SAN DIEGO – The overall rate of hospital-acquired pneumonia following cervical spinal cord injury is about 20%, results from a study of national data demonstrated.

“Cervical spinal cord injury patients are at an increased risk for the development of hospital-acquired pneumonia,” lead study author Dr. Pablo J. Diaz-Collado said in an interview after the annual meeting of the Cervical Spine Research Society.

“Complete cord injuries, longer length of stay, ICU stay and ventilation time lead to significantly increased risk of HAP, which then leads to poor inpatient outcomes,” he said. “It is of crucial importance to keep these risk factors in mind when treating patients with cervical spinal cord injuries. There is a need to optimize the management protocols for these patients to help prevent the development of HAPs.”

Dr. Diaz-Collado, an orthopedic surgery resident at Yale–New Haven (Conn.) Hospital, and his associates identified 5,198 cervical spinal cord injury patients in the 2011 and 2012 National Trauma Data Bank (NTDB) to analyze risk factors for the development of HAP and inpatient outcomes in this population. They used multivariate logistic regression to identify independent associations of various risk factors with the occurrence of HAP.

The researchers found that the overall incidence of HAP among cervical spinal cord injury patients was 20.5%, which amounted to 1,065 patients. Factors independently associated with HAP were complete spinal cord injuries (compared to central cord injuries; OR 1.44; P = .009); longer inpatient length of stay (OR 3.08 for a stay that lasted 7-13 days, OR 10.21 for 21-27 days, and OR 14.89 for 35 days or more; P = .001 or less for all associations); longer ICU stay (OR 2.86 for a stay that lasted 9-11 days, OR 3.05 for 12-14 days, and OR 2.94 for 15 days or more; P less than .001 for all associations), and longer time on mechanical ventilation (OR 2.68 for ventilation that lasted 3-6 days, OR 3.76 for 7-13 days, OR 3.98 for 14-20 days, and OR 3.99 for 21 days or more; P less than .001 for all associations).

After the researchers controlled for all other risk factors, including patient comorbidities, Injury Severity Score, and other inpatient complications, HAP was associated with increased odds of death (OR 1.60; P = .005), inpatient adverse events (OR 1.65; P less than .001), discharge to an extended-care facility (OR 1.93; P = .001), and longer length of stay (a mean of an additional 10.93 days; P less than .001).

Dr. Diaz-Collado acknowledged that the study is “limited by the quality of the data entry. In addition, the database does not include classifications of fractures, and thus stratification of the analysis in terms of the different kinds of fractures in the cervical spine is not possible. Finally, procedural codes are less accurate and thus including whether or not patients underwent a surgical intervention is less reliable.”

Dr. Diaz-Collado reported having no financial disclosures.

dbrunk@frontlinemedcom.com

SAN DIEGO – The overall rate of hospital-acquired pneumonia following cervical spinal cord injury is about 20%, results from a study of national data demonstrated.

“Cervical spinal cord injury patients are at an increased risk for the development of hospital-acquired pneumonia,” lead study author Dr. Pablo J. Diaz-Collado said in an interview after the annual meeting of the Cervical Spine Research Society.

“Complete cord injuries, longer length of stay, ICU stay and ventilation time lead to significantly increased risk of HAP, which then leads to poor inpatient outcomes,” he said. “It is of crucial importance to keep these risk factors in mind when treating patients with cervical spinal cord injuries. There is a need to optimize the management protocols for these patients to help prevent the development of HAPs.”

Dr. Diaz-Collado, an orthopedic surgery resident at Yale–New Haven (Conn.) Hospital, and his associates identified 5,198 cervical spinal cord injury patients in the 2011 and 2012 National Trauma Data Bank (NTDB) to analyze risk factors for the development of HAP and inpatient outcomes in this population. They used multivariate logistic regression to identify independent associations of various risk factors with the occurrence of HAP.

The researchers found that the overall incidence of HAP among cervical spinal cord injury patients was 20.5%, which amounted to 1,065 patients. Factors independently associated with HAP were complete spinal cord injuries (compared to central cord injuries; OR 1.44; P = .009); longer inpatient length of stay (OR 3.08 for a stay that lasted 7-13 days, OR 10.21 for 21-27 days, and OR 14.89 for 35 days or more; P = .001 or less for all associations); longer ICU stay (OR 2.86 for a stay that lasted 9-11 days, OR 3.05 for 12-14 days, and OR 2.94 for 15 days or more; P less than .001 for all associations), and longer time on mechanical ventilation (OR 2.68 for ventilation that lasted 3-6 days, OR 3.76 for 7-13 days, OR 3.98 for 14-20 days, and OR 3.99 for 21 days or more; P less than .001 for all associations).

After the researchers controlled for all other risk factors, including patient comorbidities, Injury Severity Score, and other inpatient complications, HAP was associated with increased odds of death (OR 1.60; P = .005), inpatient adverse events (OR 1.65; P less than .001), discharge to an extended-care facility (OR 1.93; P = .001), and longer length of stay (a mean of an additional 10.93 days; P less than .001).

Dr. Diaz-Collado acknowledged that the study is “limited by the quality of the data entry. In addition, the database does not include classifications of fractures, and thus stratification of the analysis in terms of the different kinds of fractures in the cervical spine is not possible. Finally, procedural codes are less accurate and thus including whether or not patients underwent a surgical intervention is less reliable.”

Dr. Diaz-Collado reported having no financial disclosures.

dbrunk@frontlinemedcom.com

SAN DIEGO – The overall rate of hospital-acquired pneumonia following cervical spinal cord injury is about 20%, results from a study of national data demonstrated.

“Cervical spinal cord injury patients are at an increased risk for the development of hospital-acquired pneumonia,” lead study author Dr. Pablo J. Diaz-Collado said in an interview after the annual meeting of the Cervical Spine Research Society.

“Complete cord injuries, longer length of stay, ICU stay and ventilation time lead to significantly increased risk of HAP, which then leads to poor inpatient outcomes,” he said. “It is of crucial importance to keep these risk factors in mind when treating patients with cervical spinal cord injuries. There is a need to optimize the management protocols for these patients to help prevent the development of HAPs.”

Dr. Diaz-Collado, an orthopedic surgery resident at Yale–New Haven (Conn.) Hospital, and his associates identified 5,198 cervical spinal cord injury patients in the 2011 and 2012 National Trauma Data Bank (NTDB) to analyze risk factors for the development of HAP and inpatient outcomes in this population. They used multivariate logistic regression to identify independent associations of various risk factors with the occurrence of HAP.

The researchers found that the overall incidence of HAP among cervical spinal cord injury patients was 20.5%, which amounted to 1,065 patients. Factors independently associated with HAP were complete spinal cord injuries (compared to central cord injuries; OR 1.44; P = .009); longer inpatient length of stay (OR 3.08 for a stay that lasted 7-13 days, OR 10.21 for 21-27 days, and OR 14.89 for 35 days or more; P = .001 or less for all associations); longer ICU stay (OR 2.86 for a stay that lasted 9-11 days, OR 3.05 for 12-14 days, and OR 2.94 for 15 days or more; P less than .001 for all associations), and longer time on mechanical ventilation (OR 2.68 for ventilation that lasted 3-6 days, OR 3.76 for 7-13 days, OR 3.98 for 14-20 days, and OR 3.99 for 21 days or more; P less than .001 for all associations).

After the researchers controlled for all other risk factors, including patient comorbidities, Injury Severity Score, and other inpatient complications, HAP was associated with increased odds of death (OR 1.60; P = .005), inpatient adverse events (OR 1.65; P less than .001), discharge to an extended-care facility (OR 1.93; P = .001), and longer length of stay (a mean of an additional 10.93 days; P less than .001).

Dr. Diaz-Collado acknowledged that the study is “limited by the quality of the data entry. In addition, the database does not include classifications of fractures, and thus stratification of the analysis in terms of the different kinds of fractures in the cervical spine is not possible. Finally, procedural codes are less accurate and thus including whether or not patients underwent a surgical intervention is less reliable.”

Dr. Diaz-Collado reported having no financial disclosures.

dbrunk@frontlinemedcom.com

Sagar Lonial, MD

© ASCO/Todd Buchanan

Single-agent daratumumab has exhibited “encouraging efficacy” and a “favorable safety profile” in patients with heavily pretreated and refractory multiple myeloma (MM), according to investigators from the phase 2 SIRIUS trial.

The drug produced an overall response rate of 30% in MM patients who had received 3 or more prior lines of therapy. The median progression-free survival was close to 4 months, and the median overall survival was nearly 18 months.

Thirty percent of patients had treatment-emergent serious adverse events (AEs), and 23% had grade 3 or 4 treatment-emergent serious AEs.

“This represents the first single-agent activity we have for a monoclonal antibody in treating multiple myeloma,” said study author Sagar Lonial, MD, of Emory University School of Medicine in Atlanta, Georgia.

“The future hope for daratumumab is in our ability to bring this active agent to earlier lines of therapy and combine it with drugs where you may get synergy.”

Dr Lonial and his colleagues reported results from the ongoing SIRIUS trial in The Lancet. Results from the trial were previously presented at the 2015 ASCO Annual Meeting. The research was funded by Janssen Research & Development, the company developing daratumumab.

In part 1 of the trial, 34 MM patients were randomized to receive either 8 mg/kg of daratumumab once every 4 weeks or 16 mg/kg once a week for 8 weeks, then once every 2 weeks for 16 weeks and once every 4 weeks after that, until disease progression or unacceptable toxicity.

In part 2, an additional 90 MM patients were enrolled and received 16 mg/kg of daratumumab on the same dosing schedule as in part 1.

Dr Lonial and his colleagues reported results for all patients who received 16 mg/kg of daratumumab. At the first interim analysis, the 8 mg/kg arm did not meet the criteria for expansion because the overall response rate was 11.1%.

The 106 patients who received the 16 mg/kg dose of daratumumab had received a median of 5 prior lines of therapy, including a proteasome inhibitor and an immunomodulatory drug. Ninety-seven percent of these patients were refractory to their last line of therapy, and 95% were refractory to both a proteasome inhibitor and an immunomodulatory drug.

Response and survival

According to an independent review committee, 29.2% of patients responded to daratumumab. Eighteen patients had a partial response, 10 had a very good partial response, and 3 had a stringent complete response.

The median duration of response was 7.4 months, and the median time to first response was 1 month.

The median overall survival was 17.5 months, and the 12-month overall survival was 64.8%. The median progression-free survival was 3.7 months.

Safety

The most common AEs were fatigue (40%), anemia (33%), nausea (29%), thrombocytopenia (25%), neutropenia (23%), back pain (22%), and cough (21%). Thirty percent of patients experienced serious AEs, and 23% had serious grade 3/4 AEs.

Infusion-related reactions were reported in 42% of patients and were predominantly grade 1 or 2 (5% grade 3; no grade 4). The most common infusion-related reactions were nasal congestion (12%), throat irritation (7%), cough (6%), dyspnea (6%), chills (6%), and vomiting (6%)—all of which were treated with standard of care and slower infusion rates.

None of the patients discontinued daratumumab because of drug-related treatment-emergent AEs, infusion-related reactions, or death. However, 5% of patients discontinued treatment because of treatment-emergent AEs—2 cases of progressive disease and 1 case each of H1N1 influenza, hypercalcemia, and spinal cord compression.

Twenty-nine percent of patients died after treatment—27% due to progressive disease and 2% due to AEs. The 2 AEs were cardiorespiratory failure secondary to H1N1 influenza complications and general health deterioration secondary to complications of aspiration pneumonia.

Sagar Lonial, MD

© ASCO/Todd Buchanan

Single-agent daratumumab has exhibited “encouraging efficacy” and a “favorable safety profile” in patients with heavily pretreated and refractory multiple myeloma (MM), according to investigators from the phase 2 SIRIUS trial.

The drug produced an overall response rate of 30% in MM patients who had received 3 or more prior lines of therapy. The median progression-free survival was close to 4 months, and the median overall survival was nearly 18 months.

Thirty percent of patients had treatment-emergent serious adverse events (AEs), and 23% had grade 3 or 4 treatment-emergent serious AEs.

“This represents the first single-agent activity we have for a monoclonal antibody in treating multiple myeloma,” said study author Sagar Lonial, MD, of Emory University School of Medicine in Atlanta, Georgia.

“The future hope for daratumumab is in our ability to bring this active agent to earlier lines of therapy and combine it with drugs where you may get synergy.”

Dr Lonial and his colleagues reported results from the ongoing SIRIUS trial in The Lancet. Results from the trial were previously presented at the 2015 ASCO Annual Meeting. The research was funded by Janssen Research & Development, the company developing daratumumab.

In part 1 of the trial, 34 MM patients were randomized to receive either 8 mg/kg of daratumumab once every 4 weeks or 16 mg/kg once a week for 8 weeks, then once every 2 weeks for 16 weeks and once every 4 weeks after that, until disease progression or unacceptable toxicity.

In part 2, an additional 90 MM patients were enrolled and received 16 mg/kg of daratumumab on the same dosing schedule as in part 1.

Dr Lonial and his colleagues reported results for all patients who received 16 mg/kg of daratumumab. At the first interim analysis, the 8 mg/kg arm did not meet the criteria for expansion because the overall response rate was 11.1%.

The 106 patients who received the 16 mg/kg dose of daratumumab had received a median of 5 prior lines of therapy, including a proteasome inhibitor and an immunomodulatory drug. Ninety-seven percent of these patients were refractory to their last line of therapy, and 95% were refractory to both a proteasome inhibitor and an immunomodulatory drug.

Response and survival

According to an independent review committee, 29.2% of patients responded to daratumumab. Eighteen patients had a partial response, 10 had a very good partial response, and 3 had a stringent complete response.

The median duration of response was 7.4 months, and the median time to first response was 1 month.

The median overall survival was 17.5 months, and the 12-month overall survival was 64.8%. The median progression-free survival was 3.7 months.

Safety

The most common AEs were fatigue (40%), anemia (33%), nausea (29%), thrombocytopenia (25%), neutropenia (23%), back pain (22%), and cough (21%). Thirty percent of patients experienced serious AEs, and 23% had serious grade 3/4 AEs.

Infusion-related reactions were reported in 42% of patients and were predominantly grade 1 or 2 (5% grade 3; no grade 4). The most common infusion-related reactions were nasal congestion (12%), throat irritation (7%), cough (6%), dyspnea (6%), chills (6%), and vomiting (6%)—all of which were treated with standard of care and slower infusion rates.

None of the patients discontinued daratumumab because of drug-related treatment-emergent AEs, infusion-related reactions, or death. However, 5% of patients discontinued treatment because of treatment-emergent AEs—2 cases of progressive disease and 1 case each of H1N1 influenza, hypercalcemia, and spinal cord compression.

Twenty-nine percent of patients died after treatment—27% due to progressive disease and 2% due to AEs. The 2 AEs were cardiorespiratory failure secondary to H1N1 influenza complications and general health deterioration secondary to complications of aspiration pneumonia.

Sagar Lonial, MD

© ASCO/Todd Buchanan

Single-agent daratumumab has exhibited “encouraging efficacy” and a “favorable safety profile” in patients with heavily pretreated and refractory multiple myeloma (MM), according to investigators from the phase 2 SIRIUS trial.

The drug produced an overall response rate of 30% in MM patients who had received 3 or more prior lines of therapy. The median progression-free survival was close to 4 months, and the median overall survival was nearly 18 months.

Thirty percent of patients had treatment-emergent serious adverse events (AEs), and 23% had grade 3 or 4 treatment-emergent serious AEs.

“This represents the first single-agent activity we have for a monoclonal antibody in treating multiple myeloma,” said study author Sagar Lonial, MD, of Emory University School of Medicine in Atlanta, Georgia.

“The future hope for daratumumab is in our ability to bring this active agent to earlier lines of therapy and combine it with drugs where you may get synergy.”

Dr Lonial and his colleagues reported results from the ongoing SIRIUS trial in The Lancet. Results from the trial were previously presented at the 2015 ASCO Annual Meeting. The research was funded by Janssen Research & Development, the company developing daratumumab.

In part 1 of the trial, 34 MM patients were randomized to receive either 8 mg/kg of daratumumab once every 4 weeks or 16 mg/kg once a week for 8 weeks, then once every 2 weeks for 16 weeks and once every 4 weeks after that, until disease progression or unacceptable toxicity.

In part 2, an additional 90 MM patients were enrolled and received 16 mg/kg of daratumumab on the same dosing schedule as in part 1.

Dr Lonial and his colleagues reported results for all patients who received 16 mg/kg of daratumumab. At the first interim analysis, the 8 mg/kg arm did not meet the criteria for expansion because the overall response rate was 11.1%.

The 106 patients who received the 16 mg/kg dose of daratumumab had received a median of 5 prior lines of therapy, including a proteasome inhibitor and an immunomodulatory drug. Ninety-seven percent of these patients were refractory to their last line of therapy, and 95% were refractory to both a proteasome inhibitor and an immunomodulatory drug.

Response and survival

According to an independent review committee, 29.2% of patients responded to daratumumab. Eighteen patients had a partial response, 10 had a very good partial response, and 3 had a stringent complete response.

The median duration of response was 7.4 months, and the median time to first response was 1 month.

The median overall survival was 17.5 months, and the 12-month overall survival was 64.8%. The median progression-free survival was 3.7 months.

Safety

The most common AEs were fatigue (40%), anemia (33%), nausea (29%), thrombocytopenia (25%), neutropenia (23%), back pain (22%), and cough (21%). Thirty percent of patients experienced serious AEs, and 23% had serious grade 3/4 AEs.

Infusion-related reactions were reported in 42% of patients and were predominantly grade 1 or 2 (5% grade 3; no grade 4). The most common infusion-related reactions were nasal congestion (12%), throat irritation (7%), cough (6%), dyspnea (6%), chills (6%), and vomiting (6%)—all of which were treated with standard of care and slower infusion rates.

None of the patients discontinued daratumumab because of drug-related treatment-emergent AEs, infusion-related reactions, or death. However, 5% of patients discontinued treatment because of treatment-emergent AEs—2 cases of progressive disease and 1 case each of H1N1 influenza, hypercalcemia, and spinal cord compression.

Twenty-nine percent of patients died after treatment—27% due to progressive disease and 2% due to AEs. The 2 AEs were cardiorespiratory failure secondary to H1N1 influenza complications and general health deterioration secondary to complications of aspiration pneumonia.

Woman sunbathing

People residing at higher latitudes, with lower exposure to sunlight/ultraviolet B (UVB) rays, have at least a 2-fold greater risk of developing leukemia than equatorial populations, according to research published in PLOS ONE.

“These results suggest that much of the burden of leukemia worldwide is due to the epidemic of vitamin D deficiency we are experiencing in winter in populations distant from the equator,” said Cedric Garland, DrPH, of the University of California San Diego in La Jolla, California.

“People who live in areas with low solar ultraviolet B exposure tend to have low levels of vitamin D metabolites in their blood. These low levels place them at high risk of certain cancers, including leukemia.”

Dr Garland and his colleagues analyzed age-adjusted incidence rates of leukemia in 172 countries and compared that information with cloud cover data from the International Satellite Cloud Climatology Project.

The team found that leukemia rates were highest in countries relatively closer to the poles, such as Australia, New Zealand, Chile, Ireland, Canada, and the United States.

And leukemia rates were lowest in countries closer to the equator, such as Bolivia, Samoa, Madagascar, and Nigeria.

The researchers also discovered that leukemia incidence was inversely associated with cloud-adjusted UVB irradiance in males (P≤0.01) and females (P≤0.01) in both hemispheres.

The association persisted in males (P≤0.05) and females (P≤0.01) after the team controlled for elevation and life expectancy.

The researchers said it’s plausible that the association is due to vitamin D deficiency.

This study follows similar investigations by Dr Garland and his colleagues in which they looked at other cancers, including breast, colon, pancreas, bladder, and multiple myeloma. In each study, the team found that reduced UVB radiation exposure and lower vitamin D levels were associated with higher risks of cancer.

“These studies do not necessarily provide final evidence,” Dr Garland said, “but they have been helpful in the past in identifying associations that have helped minimize cancer risk.”

Woman sunbathing

People residing at higher latitudes, with lower exposure to sunlight/ultraviolet B (UVB) rays, have at least a 2-fold greater risk of developing leukemia than equatorial populations, according to research published in PLOS ONE.

“These results suggest that much of the burden of leukemia worldwide is due to the epidemic of vitamin D deficiency we are experiencing in winter in populations distant from the equator,” said Cedric Garland, DrPH, of the University of California San Diego in La Jolla, California.

“People who live in areas with low solar ultraviolet B exposure tend to have low levels of vitamin D metabolites in their blood. These low levels place them at high risk of certain cancers, including leukemia.”

Dr Garland and his colleagues analyzed age-adjusted incidence rates of leukemia in 172 countries and compared that information with cloud cover data from the International Satellite Cloud Climatology Project.

The team found that leukemia rates were highest in countries relatively closer to the poles, such as Australia, New Zealand, Chile, Ireland, Canada, and the United States.

And leukemia rates were lowest in countries closer to the equator, such as Bolivia, Samoa, Madagascar, and Nigeria.

The researchers also discovered that leukemia incidence was inversely associated with cloud-adjusted UVB irradiance in males (P≤0.01) and females (P≤0.01) in both hemispheres.

The association persisted in males (P≤0.05) and females (P≤0.01) after the team controlled for elevation and life expectancy.

The researchers said it’s plausible that the association is due to vitamin D deficiency.

This study follows similar investigations by Dr Garland and his colleagues in which they looked at other cancers, including breast, colon, pancreas, bladder, and multiple myeloma. In each study, the team found that reduced UVB radiation exposure and lower vitamin D levels were associated with higher risks of cancer.

“These studies do not necessarily provide final evidence,” Dr Garland said, “but they have been helpful in the past in identifying associations that have helped minimize cancer risk.”

Woman sunbathing

People residing at higher latitudes, with lower exposure to sunlight/ultraviolet B (UVB) rays, have at least a 2-fold greater risk of developing leukemia than equatorial populations, according to research published in PLOS ONE.

“These results suggest that much of the burden of leukemia worldwide is due to the epidemic of vitamin D deficiency we are experiencing in winter in populations distant from the equator,” said Cedric Garland, DrPH, of the University of California San Diego in La Jolla, California.

“People who live in areas with low solar ultraviolet B exposure tend to have low levels of vitamin D metabolites in their blood. These low levels place them at high risk of certain cancers, including leukemia.”

Dr Garland and his colleagues analyzed age-adjusted incidence rates of leukemia in 172 countries and compared that information with cloud cover data from the International Satellite Cloud Climatology Project.

The team found that leukemia rates were highest in countries relatively closer to the poles, such as Australia, New Zealand, Chile, Ireland, Canada, and the United States.

And leukemia rates were lowest in countries closer to the equator, such as Bolivia, Samoa, Madagascar, and Nigeria.

The researchers also discovered that leukemia incidence was inversely associated with cloud-adjusted UVB irradiance in males (P≤0.01) and females (P≤0.01) in both hemispheres.

The association persisted in males (P≤0.05) and females (P≤0.01) after the team controlled for elevation and life expectancy.

The researchers said it’s plausible that the association is due to vitamin D deficiency.

This study follows similar investigations by Dr Garland and his colleagues in which they looked at other cancers, including breast, colon, pancreas, bladder, and multiple myeloma. In each study, the team found that reduced UVB radiation exposure and lower vitamin D levels were associated with higher risks of cancer.

“These studies do not necessarily provide final evidence,” Dr Garland said, “but they have been helpful in the past in identifying associations that have helped minimize cancer risk.”

Micrograph showing CTCL

New research indicates that toxins in Staphylococcus bacteria help malignant cells gain control over healthy cells in patients with cutaneous T-cell lymphoma (CTCL).

Investigators found that staphylococcal enterotoxin-A (SEA) induces STAT3 activation and IL-17 expression in malignant T cells via engagement of non-malignant CD4 T cells.

As STAT3 activation has been implicated in CTCL pathogenesis, the discovery suggests bacterial toxins play a key role in activating an oncogenic pathway in CTCL.

“We have gained important insight into the processes that activate cancer cells and make them grow,” said Niels Oedum, MD, of the University of Copenhagen in Denmark.

“[CTCL] patients’ frequent bacterial infections might not be a mere side effect of the disease. On the contrary, toxins in the bacteria actually ‘benefit’ cancer cells. Our next step is examining whether combatting infections can slow down the growth of cancer cells and thus stop the disease.”

Dr Oedum and his colleagues described their research in Blood.

The investigators knew that, in CTCL, CD4 T cells become malignant and turn parasitic on the rest of the immune system. In addition to using healthy cells to do their work for them, the malignant cells slowly destroy the skin’s immune defense mechanism.

The team’s new discoveries indicate that bacterial toxins in some patients enable malignant cells to send off signals that obstruct and change the immune defense mechanism, which would otherwise fight the malignant cells. What was believed to be an overly active immune defense mechanism could, in other words, turn out to be a malignant infection brought on by bacteria, which only worsens the disease.

Dr Oedum and his colleagues found that SEA-positive bacteria isolatated from the skin of CTCL patients stimulated activation of STAT3 and upregulation of IL-17 in malignant and non-malignant T cells.

Malignant T cells expressing an SEA non-responsive T-cell receptor V beta chain did not respond to SEA when cultured alone but exhibited STAT3 activation and IL-17 expression in co-cultures with SEA-responsive, non-malignant T cells.

The investigators found evidence to suggest the response is induced via IL-2Rg cytokines and a JAK3-dependent pathway in malignant T cells. The JAK3 inhibitor tofacitinib inhibited SEA-induced IL-17 production in co-cultures of malignant and non-malignant T cells.

Dr Oedum and his colleagues plan to continue their work investigating how bacteria might affect the balance between the immune defense mechanism and the disease in patients with CTCL.

In the long-term, the investigators’ aim is to understand how bacteria and their toxins can worsen CTCL, knowledge that may be used to develop new targeted treatments.

As only some of the bacteria produce toxins, the team said it will also be important to develop methods to determine which patients may benefit from treatment with antibiotics.

Micrograph showing CTCL

New research indicates that toxins in Staphylococcus bacteria help malignant cells gain control over healthy cells in patients with cutaneous T-cell lymphoma (CTCL).

Investigators found that staphylococcal enterotoxin-A (SEA) induces STAT3 activation and IL-17 expression in malignant T cells via engagement of non-malignant CD4 T cells.

As STAT3 activation has been implicated in CTCL pathogenesis, the discovery suggests bacterial toxins play a key role in activating an oncogenic pathway in CTCL.

“We have gained important insight into the processes that activate cancer cells and make them grow,” said Niels Oedum, MD, of the University of Copenhagen in Denmark.

“[CTCL] patients’ frequent bacterial infections might not be a mere side effect of the disease. On the contrary, toxins in the bacteria actually ‘benefit’ cancer cells. Our next step is examining whether combatting infections can slow down the growth of cancer cells and thus stop the disease.”

Dr Oedum and his colleagues described their research in Blood.

The investigators knew that, in CTCL, CD4 T cells become malignant and turn parasitic on the rest of the immune system. In addition to using healthy cells to do their work for them, the malignant cells slowly destroy the skin’s immune defense mechanism.

The team’s new discoveries indicate that bacterial toxins in some patients enable malignant cells to send off signals that obstruct and change the immune defense mechanism, which would otherwise fight the malignant cells. What was believed to be an overly active immune defense mechanism could, in other words, turn out to be a malignant infection brought on by bacteria, which only worsens the disease.

Dr Oedum and his colleagues found that SEA-positive bacteria isolatated from the skin of CTCL patients stimulated activation of STAT3 and upregulation of IL-17 in malignant and non-malignant T cells.

Malignant T cells expressing an SEA non-responsive T-cell receptor V beta chain did not respond to SEA when cultured alone but exhibited STAT3 activation and IL-17 expression in co-cultures with SEA-responsive, non-malignant T cells.

The investigators found evidence to suggest the response is induced via IL-2Rg cytokines and a JAK3-dependent pathway in malignant T cells. The JAK3 inhibitor tofacitinib inhibited SEA-induced IL-17 production in co-cultures of malignant and non-malignant T cells.

Dr Oedum and his colleagues plan to continue their work investigating how bacteria might affect the balance between the immune defense mechanism and the disease in patients with CTCL.

In the long-term, the investigators’ aim is to understand how bacteria and their toxins can worsen CTCL, knowledge that may be used to develop new targeted treatments.

As only some of the bacteria produce toxins, the team said it will also be important to develop methods to determine which patients may benefit from treatment with antibiotics.

Micrograph showing CTCL

New research indicates that toxins in Staphylococcus bacteria help malignant cells gain control over healthy cells in patients with cutaneous T-cell lymphoma (CTCL).

Investigators found that staphylococcal enterotoxin-A (SEA) induces STAT3 activation and IL-17 expression in malignant T cells via engagement of non-malignant CD4 T cells.

As STAT3 activation has been implicated in CTCL pathogenesis, the discovery suggests bacterial toxins play a key role in activating an oncogenic pathway in CTCL.

“We have gained important insight into the processes that activate cancer cells and make them grow,” said Niels Oedum, MD, of the University of Copenhagen in Denmark.

“[CTCL] patients’ frequent bacterial infections might not be a mere side effect of the disease. On the contrary, toxins in the bacteria actually ‘benefit’ cancer cells. Our next step is examining whether combatting infections can slow down the growth of cancer cells and thus stop the disease.”

Dr Oedum and his colleagues described their research in Blood.

The investigators knew that, in CTCL, CD4 T cells become malignant and turn parasitic on the rest of the immune system. In addition to using healthy cells to do their work for them, the malignant cells slowly destroy the skin’s immune defense mechanism.

The team’s new discoveries indicate that bacterial toxins in some patients enable malignant cells to send off signals that obstruct and change the immune defense mechanism, which would otherwise fight the malignant cells. What was believed to be an overly active immune defense mechanism could, in other words, turn out to be a malignant infection brought on by bacteria, which only worsens the disease.

Dr Oedum and his colleagues found that SEA-positive bacteria isolatated from the skin of CTCL patients stimulated activation of STAT3 and upregulation of IL-17 in malignant and non-malignant T cells.

Malignant T cells expressing an SEA non-responsive T-cell receptor V beta chain did not respond to SEA when cultured alone but exhibited STAT3 activation and IL-17 expression in co-cultures with SEA-responsive, non-malignant T cells.

The investigators found evidence to suggest the response is induced via IL-2Rg cytokines and a JAK3-dependent pathway in malignant T cells. The JAK3 inhibitor tofacitinib inhibited SEA-induced IL-17 production in co-cultures of malignant and non-malignant T cells.

Dr Oedum and his colleagues plan to continue their work investigating how bacteria might affect the balance between the immune defense mechanism and the disease in patients with CTCL.

In the long-term, the investigators’ aim is to understand how bacteria and their toxins can worsen CTCL, knowledge that may be used to develop new targeted treatments.

As only some of the bacteria produce toxins, the team said it will also be important to develop methods to determine which patients may benefit from treatment with antibiotics.

Doctor and patient in hospital

Photo courtesy of the CDC

Patients with incurable cancer and multiple comorbidities who consulted with a palliative care team within 2 days of hospitalization had significant savings in hospital costs, according to a new study.

The study also showed that the higher number of comorbidities a patient had, the greater the reduction in direct hospital costs with early palliative care as opposed to standard care.

Previous studies have shown a link between palliative care and lower costs, but this is the first to examine whether the effect of palliative care consultation varies by the number of co-existing chronic conditions.

“We already know that coordinated, patient-centered palliative care improves care quality, enhances survival, and reduces costs for persons with cancer,” said R. Sean Morrison, MD, of the Icahn School of Medicine at Mount Sinai in New York, New York.

“Our latest research now shows the strong association between cost and the number of co-occurring conditions. Among patients with advanced cancer and other serious illnesses, aggressive treatments are often inconsistent with patients’ wishes and are associated with worse quality of life compared to other treatments. It is imperative that policymakers act to expand access to palliative care.”

Dr Morrison and his colleagues described their research in Health Affairs.

The study included 906 patients with advanced cancer and multiple comorbidities who were treated at 6 hospitals. One hundred and ninety-three patients were seen by a palliative care team within 2 days of hospitalization, while the remaining 713 patients received usual care.

Patients from the palliative care group had significantly lower total direct hospital costs if they had multimorbidity. For patients with a comorbidity score of 0–1, the estimated mean treatment effect was not significant.

However, patients with a comorbidity score of 2–3 had a 22% reduction in costs, or a reduction of $2321. Patients with a score of 4 or higher had a cost reduction of 32%, or $3515.

“The fact that we found greater cost savings for cancer patients with more comorbidities than for those with fewer comorbidities raises the question of whether similar results would be observed in patients with other serious illnesses and multimorbidity,” said Peter May, of Trinity College Dublin in Ireland.

“Future research is also needed to determine when in the course of illness palliative care is most cost-effective.”

Doctor and patient in hospital

Photo courtesy of the CDC

Patients with incurable cancer and multiple comorbidities who consulted with a palliative care team within 2 days of hospitalization had significant savings in hospital costs, according to a new study.

The study also showed that the higher number of comorbidities a patient had, the greater the reduction in direct hospital costs with early palliative care as opposed to standard care.

Previous studies have shown a link between palliative care and lower costs, but this is the first to examine whether the effect of palliative care consultation varies by the number of co-existing chronic conditions.

“We already know that coordinated, patient-centered palliative care improves care quality, enhances survival, and reduces costs for persons with cancer,” said R. Sean Morrison, MD, of the Icahn School of Medicine at Mount Sinai in New York, New York.

“Our latest research now shows the strong association between cost and the number of co-occurring conditions. Among patients with advanced cancer and other serious illnesses, aggressive treatments are often inconsistent with patients’ wishes and are associated with worse quality of life compared to other treatments. It is imperative that policymakers act to expand access to palliative care.”

Dr Morrison and his colleagues described their research in Health Affairs.

The study included 906 patients with advanced cancer and multiple comorbidities who were treated at 6 hospitals. One hundred and ninety-three patients were seen by a palliative care team within 2 days of hospitalization, while the remaining 713 patients received usual care.

Patients from the palliative care group had significantly lower total direct hospital costs if they had multimorbidity. For patients with a comorbidity score of 0–1, the estimated mean treatment effect was not significant.

However, patients with a comorbidity score of 2–3 had a 22% reduction in costs, or a reduction of $2321. Patients with a score of 4 or higher had a cost reduction of 32%, or $3515.

“The fact that we found greater cost savings for cancer patients with more comorbidities than for those with fewer comorbidities raises the question of whether similar results would be observed in patients with other serious illnesses and multimorbidity,” said Peter May, of Trinity College Dublin in Ireland.

“Future research is also needed to determine when in the course of illness palliative care is most cost-effective.”

Doctor and patient in hospital

Photo courtesy of the CDC

Patients with incurable cancer and multiple comorbidities who consulted with a palliative care team within 2 days of hospitalization had significant savings in hospital costs, according to a new study.

The study also showed that the higher number of comorbidities a patient had, the greater the reduction in direct hospital costs with early palliative care as opposed to standard care.

Previous studies have shown a link between palliative care and lower costs, but this is the first to examine whether the effect of palliative care consultation varies by the number of co-existing chronic conditions.

“We already know that coordinated, patient-centered palliative care improves care quality, enhances survival, and reduces costs for persons with cancer,” said R. Sean Morrison, MD, of the Icahn School of Medicine at Mount Sinai in New York, New York.

“Our latest research now shows the strong association between cost and the number of co-occurring conditions. Among patients with advanced cancer and other serious illnesses, aggressive treatments are often inconsistent with patients’ wishes and are associated with worse quality of life compared to other treatments. It is imperative that policymakers act to expand access to palliative care.”

Dr Morrison and his colleagues described their research in Health Affairs.

The study included 906 patients with advanced cancer and multiple comorbidities who were treated at 6 hospitals. One hundred and ninety-three patients were seen by a palliative care team within 2 days of hospitalization, while the remaining 713 patients received usual care.

Patients from the palliative care group had significantly lower total direct hospital costs if they had multimorbidity. For patients with a comorbidity score of 0–1, the estimated mean treatment effect was not significant.

However, patients with a comorbidity score of 2–3 had a 22% reduction in costs, or a reduction of $2321. Patients with a score of 4 or higher had a cost reduction of 32%, or $3515.

“The fact that we found greater cost savings for cancer patients with more comorbidities than for those with fewer comorbidities raises the question of whether similar results would be observed in patients with other serious illnesses and multimorbidity,” said Peter May, of Trinity College Dublin in Ireland.

“Future research is also needed to determine when in the course of illness palliative care is most cost-effective.”

This is the first installment of a five-part monthly series that will discuss the pathologic, genomic, and clinical factors that contribute to the racial survival disparity in breast cancer. The series, which is adapted from an article that originally appeared in CA: A Cancer Journal for Clinicians,¹ a journal of the American Cancer Society, will also review exciting and innovative interventions to close this survival gap. This month’s column reviews the scope of this important health care issue.

The National Cancer Institute’s (NCI) Surveillance, Epidemiology, and End Results Program (SEER) has estimated that 231,840 new cases of female breast cancer will be diagnosed in 2015, representing 14% of all new cancer cases among women. The NCI also has estimated 40,290 deaths from breast cancer, representing 6.8% of all cancer deaths among women.² Breast cancer is the second leading cause of cancer death among women after lung cancer. It is well known that there has historically been a significant racial divide in breast cancer incidence (rate of new occurrences of breast cancer) and mortality (death) rates. Caucasian women were more likely to be diagnosed with breast cancer, but African American women were more likely to die from it.

However, in a recently released study by DeSantis et al. this incidence trend no longer holds, and in 2012 there was a convergence of breast cancer incidence rates at 135 cases per 100,000 women for both Caucasian and African American women.³ In addition, this recent analysis revealed that the mortality disparity between African American and Caucasian women has continued to increase, with a death rate 42% higher in African American than in Caucasian women in 2012. While overall improvements in therapy have led to a decrease in breast cancer death rates in the United States since 1990, the decreases in death rates began earlier and have been larger in proportionate terms for Caucasians than for African Americans.^4,5 According to SEER data from 1975 to 2011, Caucasian women had a 23% increase in breast cancer incidence and a 34% decrease in mortality, whereas African American women experienced a 35% increase in incidence and a 2% increase in mortality.⁶

Beyond national statistics and on a more-local level, several studies have explored regional variations in breast cancer mortality by race. One such study analyzed mortality data from the National Center for Health Statistics from 1975 to 2004.⁵ The researchers discovered that trends in breast cancer death rates varied widely by region. While breast cancer death rates in Caucasian women decreased in all 50 states, among African American women in 37 states analyzed, breast cancer death rates increased in 2 states, were level in 24 states, and decreased in only 11 states. Many of the states in which African American breast cancer death rates were level or rising were in the South and Midwest.

There are also differences in age and stage at diagnosis between African American and Caucasian women. Although the overall incidence of breast cancer has been historically higher in Caucasians, the incidence profile changes when the data are looked at by age. Among African American women with breast cancer, 33% are diagnosed at an age younger than 50 years, compared with 21.9% among Caucasian women.⁷

In women younger than 35 years, the incidence of breast cancer in African Americans is 1.4-2.0 times that of Caucasians.⁸ In addition, African American women present with more advanced-stage disease. Again, using the SEER program and examining data from 2005-2011, 62% of Caucasians had localized disease (cancer confined to the breast and potentially curable) versus 53% of African Americans. In all, 5% of Caucasians had distant disease (cancer outside the breast and treatable but not curable), compared with 9% of African Americans.⁹ A recent study in JAMA of 373,563 women with breast cancer during 2004-2011 found that African American women were less likely to be diagnosed with stage I breast cancer than were non-Hispanic white women across all age groups (non-Hispanic white women, 50.8%; African American women, 37.0%).¹⁰

The researchers examined further those women with small breast cancers (breast tumors ≤ 2 cm) and the percentages of nodal metastases (cancer in the lymph nodes) and distant metastases (cancer outside the breast) by race/ethnicity. The authors found that an African American woman with a small-sized breast tumor was more likely to present with lymph node metastases and distant metastases. Significantly, African American women were also more likely to die of breast cancer with small-sized tumors than were non-Hispanic white women.

These differences in age and stage highlight important differences in tumor biology, genomics, and patterns of care that contribute to the disparity in breast cancer survival between Caucasian and African American women. The February installment of this column will explore tumor biology – the first element in the perfect storm.

Other installments of this column can be found in the Related Content box.

 1. Daly B, Olopade OI: A perfect storm: How tumor biology, genomics, and health care delivery patterns collide to create a racial survival disparity in breast cancer and proposed interventions for change. CA Cancer J Clin. 65:221-38, 2015.

 2. National Cancer Institute. Surveillance, Epidemiology, and End Results (SEER) Program Stat fact sheets: Breast cancer. Surveillance, Epidemiology, and End Results Program. http://seer.cancer.gov/statfacts/html/breast.html. Accessed Nov. 20, 2015.

 3. DeSantis C, Fedewa S, Goding Sauer A, et al., Breast cancer statistics, 2015: Convergence of incidence rates between black and white women. CA: A Cancer Journal for Clinicians. doi: 10.3322/caac.21320

 4. DeLancey JO, Thun MJ, Jemal A, et al.: Recent trends in Black-White disparities in cancer mortality. Cancer Epidemiol Biomarkers Prev. 17:2908-12, 2008.

 5. DeSantis C, Jemal A, Ward E, et al.: Temporal trends in breast cancer mortality by state and race. Cancer Causes Control. 19:537-45, 2008.

 6. Howlander N NA, Krapcho M, et al. eds.: SEER Cancer Statistics Review, 1975-2011, 2014.

 7. Clarke CA, West DW, Edwards BK, et al.: Existing data on breast cancer in African-American women: what we know and what we need to know. Cancer. 97:211-21, 2003.

 8. Marie Swanson G, Haslam SZ, Azzouz F: Breast cancer among young African-American women: a summary of data and literature and of issues discussed during the Summit Meeting on Breast Cancer Among African American Women, Washington, DC, September 8-10, 2000. Cancer. 97:273-9, 2003.

 9. National Cancer Institute. SEER Cancer Statistics Review, 1975-2012. http://seer.cancer.gov/csr/1975_2012/results_single/sect_04_table.13.pdf. Accessed, Nov. 20, 2015.

 10. Iqbal J, Ginsburg O, Rochon PA, et al: Differences in breast cancer stage at diagnosis and cancer-specific survival by race and ethnicity in the United States. JAMA 313:165-73, 2015.

Bobby Daly, MD, MBA, is the chief fellow in the section of hematology/oncology at the University of Chicago Medicine. His clinical focus is breast and thoracic oncology, and his research focus is health services. Specifically, Dr. Daly researches disparities in oncology care delivery, oncology health care utilization, aggressive end-of-life oncology care, and oncology payment models. He received his MD and MBA from Harvard Medical School and Harvard Business School, both in Boston, and a BA in Economics and History from Stanford (Calif.) University. He was the recipient of the Dean’s Award at Harvard Medical and Business Schools.

Olufunmilayo Olopade, MD, FACP, OON, is the Walter L. Palmer Distinguished Service Professor of Medicine and Human Genetics, and director, Center for Global Health at the University of Chicago. She is adopting emerging high throughput genomic and informatics strategies to identify genetic and nongenetic risk factors for breast cancer in order to implement precision health care in diverse populations. This innovative approach has the potential to improve the quality of care and reduce costs while saving more lives.

Disclosures: Dr. Olopade serves on the Medical Advisory Board for CancerIQ. Dr. Daly serves as a director of Quadrant Holdings Corporation and receives compensation from this entity. Frontline Medical Communications is a subsidiary of Quadrant Holdings Corporation.

Published in conjunction with Susan G. Komen®.

This is the first installment of a five-part monthly series that will discuss the pathologic, genomic, and clinical factors that contribute to the racial survival disparity in breast cancer. The series, which is adapted from an article that originally appeared in CA: A Cancer Journal for Clinicians,¹ a journal of the American Cancer Society, will also review exciting and innovative interventions to close this survival gap. This month’s column reviews the scope of this important health care issue.

The National Cancer Institute’s (NCI) Surveillance, Epidemiology, and End Results Program (SEER) has estimated that 231,840 new cases of female breast cancer will be diagnosed in 2015, representing 14% of all new cancer cases among women. The NCI also has estimated 40,290 deaths from breast cancer, representing 6.8% of all cancer deaths among women.² Breast cancer is the second leading cause of cancer death among women after lung cancer. It is well known that there has historically been a significant racial divide in breast cancer incidence (rate of new occurrences of breast cancer) and mortality (death) rates. Caucasian women were more likely to be diagnosed with breast cancer, but African American women were more likely to die from it.

However, in a recently released study by DeSantis et al. this incidence trend no longer holds, and in 2012 there was a convergence of breast cancer incidence rates at 135 cases per 100,000 women for both Caucasian and African American women.³ In addition, this recent analysis revealed that the mortality disparity between African American and Caucasian women has continued to increase, with a death rate 42% higher in African American than in Caucasian women in 2012. While overall improvements in therapy have led to a decrease in breast cancer death rates in the United States since 1990, the decreases in death rates began earlier and have been larger in proportionate terms for Caucasians than for African Americans.^4,5 According to SEER data from 1975 to 2011, Caucasian women had a 23% increase in breast cancer incidence and a 34% decrease in mortality, whereas African American women experienced a 35% increase in incidence and a 2% increase in mortality.⁶

Beyond national statistics and on a more-local level, several studies have explored regional variations in breast cancer mortality by race. One such study analyzed mortality data from the National Center for Health Statistics from 1975 to 2004.⁵ The researchers discovered that trends in breast cancer death rates varied widely by region. While breast cancer death rates in Caucasian women decreased in all 50 states, among African American women in 37 states analyzed, breast cancer death rates increased in 2 states, were level in 24 states, and decreased in only 11 states. Many of the states in which African American breast cancer death rates were level or rising were in the South and Midwest.

There are also differences in age and stage at diagnosis between African American and Caucasian women. Although the overall incidence of breast cancer has been historically higher in Caucasians, the incidence profile changes when the data are looked at by age. Among African American women with breast cancer, 33% are diagnosed at an age younger than 50 years, compared with 21.9% among Caucasian women.⁷

In women younger than 35 years, the incidence of breast cancer in African Americans is 1.4-2.0 times that of Caucasians.⁸ In addition, African American women present with more advanced-stage disease. Again, using the SEER program and examining data from 2005-2011, 62% of Caucasians had localized disease (cancer confined to the breast and potentially curable) versus 53% of African Americans. In all, 5% of Caucasians had distant disease (cancer outside the breast and treatable but not curable), compared with 9% of African Americans.⁹ A recent study in JAMA of 373,563 women with breast cancer during 2004-2011 found that African American women were less likely to be diagnosed with stage I breast cancer than were non-Hispanic white women across all age groups (non-Hispanic white women, 50.8%; African American women, 37.0%).¹⁰

The researchers examined further those women with small breast cancers (breast tumors ≤ 2 cm) and the percentages of nodal metastases (cancer in the lymph nodes) and distant metastases (cancer outside the breast) by race/ethnicity. The authors found that an African American woman with a small-sized breast tumor was more likely to present with lymph node metastases and distant metastases. Significantly, African American women were also more likely to die of breast cancer with small-sized tumors than were non-Hispanic white women.

These differences in age and stage highlight important differences in tumor biology, genomics, and patterns of care that contribute to the disparity in breast cancer survival between Caucasian and African American women. The February installment of this column will explore tumor biology – the first element in the perfect storm.

Other installments of this column can be found in the Related Content box.

 1. Daly B, Olopade OI: A perfect storm: How tumor biology, genomics, and health care delivery patterns collide to create a racial survival disparity in breast cancer and proposed interventions for change. CA Cancer J Clin. 65:221-38, 2015.

 2. National Cancer Institute. Surveillance, Epidemiology, and End Results (SEER) Program Stat fact sheets: Breast cancer. Surveillance, Epidemiology, and End Results Program. http://seer.cancer.gov/statfacts/html/breast.html. Accessed Nov. 20, 2015.

 3. DeSantis C, Fedewa S, Goding Sauer A, et al., Breast cancer statistics, 2015: Convergence of incidence rates between black and white women. CA: A Cancer Journal for Clinicians. doi: 10.3322/caac.21320

 4. DeLancey JO, Thun MJ, Jemal A, et al.: Recent trends in Black-White disparities in cancer mortality. Cancer Epidemiol Biomarkers Prev. 17:2908-12, 2008.

 5. DeSantis C, Jemal A, Ward E, et al.: Temporal trends in breast cancer mortality by state and race. Cancer Causes Control. 19:537-45, 2008.

 6. Howlander N NA, Krapcho M, et al. eds.: SEER Cancer Statistics Review, 1975-2011, 2014.

 7. Clarke CA, West DW, Edwards BK, et al.: Existing data on breast cancer in African-American women: what we know and what we need to know. Cancer. 97:211-21, 2003.

 8. Marie Swanson G, Haslam SZ, Azzouz F: Breast cancer among young African-American women: a summary of data and literature and of issues discussed during the Summit Meeting on Breast Cancer Among African American Women, Washington, DC, September 8-10, 2000. Cancer. 97:273-9, 2003.

 9. National Cancer Institute. SEER Cancer Statistics Review, 1975-2012. http://seer.cancer.gov/csr/1975_2012/results_single/sect_04_table.13.pdf. Accessed, Nov. 20, 2015.

 10. Iqbal J, Ginsburg O, Rochon PA, et al: Differences in breast cancer stage at diagnosis and cancer-specific survival by race and ethnicity in the United States. JAMA 313:165-73, 2015.

Bobby Daly, MD, MBA, is the chief fellow in the section of hematology/oncology at the University of Chicago Medicine. His clinical focus is breast and thoracic oncology, and his research focus is health services. Specifically, Dr. Daly researches disparities in oncology care delivery, oncology health care utilization, aggressive end-of-life oncology care, and oncology payment models. He received his MD and MBA from Harvard Medical School and Harvard Business School, both in Boston, and a BA in Economics and History from Stanford (Calif.) University. He was the recipient of the Dean’s Award at Harvard Medical and Business Schools.

Olufunmilayo Olopade, MD, FACP, OON, is the Walter L. Palmer Distinguished Service Professor of Medicine and Human Genetics, and director, Center for Global Health at the University of Chicago. She is adopting emerging high throughput genomic and informatics strategies to identify genetic and nongenetic risk factors for breast cancer in order to implement precision health care in diverse populations. This innovative approach has the potential to improve the quality of care and reduce costs while saving more lives.

Disclosures: Dr. Olopade serves on the Medical Advisory Board for CancerIQ. Dr. Daly serves as a director of Quadrant Holdings Corporation and receives compensation from this entity. Frontline Medical Communications is a subsidiary of Quadrant Holdings Corporation.

Published in conjunction with Susan G. Komen®.

This is the first installment of a five-part monthly series that will discuss the pathologic, genomic, and clinical factors that contribute to the racial survival disparity in breast cancer. The series, which is adapted from an article that originally appeared in CA: A Cancer Journal for Clinicians,¹ a journal of the American Cancer Society, will also review exciting and innovative interventions to close this survival gap. This month’s column reviews the scope of this important health care issue.

The National Cancer Institute’s (NCI) Surveillance, Epidemiology, and End Results Program (SEER) has estimated that 231,840 new cases of female breast cancer will be diagnosed in 2015, representing 14% of all new cancer cases among women. The NCI also has estimated 40,290 deaths from breast cancer, representing 6.8% of all cancer deaths among women.² Breast cancer is the second leading cause of cancer death among women after lung cancer. It is well known that there has historically been a significant racial divide in breast cancer incidence (rate of new occurrences of breast cancer) and mortality (death) rates. Caucasian women were more likely to be diagnosed with breast cancer, but African American women were more likely to die from it.

However, in a recently released study by DeSantis et al. this incidence trend no longer holds, and in 2012 there was a convergence of breast cancer incidence rates at 135 cases per 100,000 women for both Caucasian and African American women.³ In addition, this recent analysis revealed that the mortality disparity between African American and Caucasian women has continued to increase, with a death rate 42% higher in African American than in Caucasian women in 2012. While overall improvements in therapy have led to a decrease in breast cancer death rates in the United States since 1990, the decreases in death rates began earlier and have been larger in proportionate terms for Caucasians than for African Americans.^4,5 According to SEER data from 1975 to 2011, Caucasian women had a 23% increase in breast cancer incidence and a 34% decrease in mortality, whereas African American women experienced a 35% increase in incidence and a 2% increase in mortality.⁶

Beyond national statistics and on a more-local level, several studies have explored regional variations in breast cancer mortality by race. One such study analyzed mortality data from the National Center for Health Statistics from 1975 to 2004.⁵ The researchers discovered that trends in breast cancer death rates varied widely by region. While breast cancer death rates in Caucasian women decreased in all 50 states, among African American women in 37 states analyzed, breast cancer death rates increased in 2 states, were level in 24 states, and decreased in only 11 states. Many of the states in which African American breast cancer death rates were level or rising were in the South and Midwest.

There are also differences in age and stage at diagnosis between African American and Caucasian women. Although the overall incidence of breast cancer has been historically higher in Caucasians, the incidence profile changes when the data are looked at by age. Among African American women with breast cancer, 33% are diagnosed at an age younger than 50 years, compared with 21.9% among Caucasian women.⁷

In women younger than 35 years, the incidence of breast cancer in African Americans is 1.4-2.0 times that of Caucasians.⁸ In addition, African American women present with more advanced-stage disease. Again, using the SEER program and examining data from 2005-2011, 62% of Caucasians had localized disease (cancer confined to the breast and potentially curable) versus 53% of African Americans. In all, 5% of Caucasians had distant disease (cancer outside the breast and treatable but not curable), compared with 9% of African Americans.⁹ A recent study in JAMA of 373,563 women with breast cancer during 2004-2011 found that African American women were less likely to be diagnosed with stage I breast cancer than were non-Hispanic white women across all age groups (non-Hispanic white women, 50.8%; African American women, 37.0%).¹⁰

The researchers examined further those women with small breast cancers (breast tumors ≤ 2 cm) and the percentages of nodal metastases (cancer in the lymph nodes) and distant metastases (cancer outside the breast) by race/ethnicity. The authors found that an African American woman with a small-sized breast tumor was more likely to present with lymph node metastases and distant metastases. Significantly, African American women were also more likely to die of breast cancer with small-sized tumors than were non-Hispanic white women.

These differences in age and stage highlight important differences in tumor biology, genomics, and patterns of care that contribute to the disparity in breast cancer survival between Caucasian and African American women. The February installment of this column will explore tumor biology – the first element in the perfect storm.

Other installments of this column can be found in the Related Content box.

 1. Daly B, Olopade OI: A perfect storm: How tumor biology, genomics, and health care delivery patterns collide to create a racial survival disparity in breast cancer and proposed interventions for change. CA Cancer J Clin. 65:221-38, 2015.

 2. National Cancer Institute. Surveillance, Epidemiology, and End Results (SEER) Program Stat fact sheets: Breast cancer. Surveillance, Epidemiology, and End Results Program. http://seer.cancer.gov/statfacts/html/breast.html. Accessed Nov. 20, 2015.

 3. DeSantis C, Fedewa S, Goding Sauer A, et al., Breast cancer statistics, 2015: Convergence of incidence rates between black and white women. CA: A Cancer Journal for Clinicians. doi: 10.3322/caac.21320

 4. DeLancey JO, Thun MJ, Jemal A, et al.: Recent trends in Black-White disparities in cancer mortality. Cancer Epidemiol Biomarkers Prev. 17:2908-12, 2008.

 5. DeSantis C, Jemal A, Ward E, et al.: Temporal trends in breast cancer mortality by state and race. Cancer Causes Control. 19:537-45, 2008.

 6. Howlander N NA, Krapcho M, et al. eds.: SEER Cancer Statistics Review, 1975-2011, 2014.

 7. Clarke CA, West DW, Edwards BK, et al.: Existing data on breast cancer in African-American women: what we know and what we need to know. Cancer. 97:211-21, 2003.

 8. Marie Swanson G, Haslam SZ, Azzouz F: Breast cancer among young African-American women: a summary of data and literature and of issues discussed during the Summit Meeting on Breast Cancer Among African American Women, Washington, DC, September 8-10, 2000. Cancer. 97:273-9, 2003.

 9. National Cancer Institute. SEER Cancer Statistics Review, 1975-2012. http://seer.cancer.gov/csr/1975_2012/results_single/sect_04_table.13.pdf. Accessed, Nov. 20, 2015.

 10. Iqbal J, Ginsburg O, Rochon PA, et al: Differences in breast cancer stage at diagnosis and cancer-specific survival by race and ethnicity in the United States. JAMA 313:165-73, 2015.

Bobby Daly, MD, MBA, is the chief fellow in the section of hematology/oncology at the University of Chicago Medicine. His clinical focus is breast and thoracic oncology, and his research focus is health services. Specifically, Dr. Daly researches disparities in oncology care delivery, oncology health care utilization, aggressive end-of-life oncology care, and oncology payment models. He received his MD and MBA from Harvard Medical School and Harvard Business School, both in Boston, and a BA in Economics and History from Stanford (Calif.) University. He was the recipient of the Dean’s Award at Harvard Medical and Business Schools.

Olufunmilayo Olopade, MD, FACP, OON, is the Walter L. Palmer Distinguished Service Professor of Medicine and Human Genetics, and director, Center for Global Health at the University of Chicago. She is adopting emerging high throughput genomic and informatics strategies to identify genetic and nongenetic risk factors for breast cancer in order to implement precision health care in diverse populations. This innovative approach has the potential to improve the quality of care and reduce costs while saving more lives.

Disclosures: Dr. Olopade serves on the Medical Advisory Board for CancerIQ. Dr. Daly serves as a director of Quadrant Holdings Corporation and receives compensation from this entity. Frontline Medical Communications is a subsidiary of Quadrant Holdings Corporation.

Published in conjunction with Susan G. Komen®.

Scapular dislocation, which is also termed locked scapula or scapulothoracic dislocation, is an unusual condition that can be described as extrathoracic or intrathoracic dislocation, depending on the penetration of scapula into the thoracic cavity.

There have been 3 reported cases of intrathoracic scapular dislocations in the literature,^1-3all associated with a preexisting condition (eg, sternoclavicular separation, prior rib fracture, thoracotomy for a lung transplant procedure, or surgical resection of superior ribs during breast or pulmonary tumor excisions). Three published cases of intrathoracic scapular impaction involve comminuted scapular fractures with intrathoracic impaction of the inferior fragment through intercostal space.^4-6

Here we report an intrathoracic scapular dislocation that was not associated with fracture of the scapula or predisposing factors. To our knowledge, this is the first case of pure intrathoracic dislocation. The possibility of intrathoracic scapular dislocation should be considered as part of the differential diagnosis even in patients with a negative anamnesis for predisposing factors, such as lung or chest surgery. The patient provided written informed consent for print and electronic publication of this case report.

Case Report

A 29-year-old woman presented to the emergency department after a motor vehicle accident. She had tenderness over the left shoulder and left elbow with decreased range of motion; however, motor and sensory examination of the wrist and fingers were normal. No distal neurovascular deficit was noted.

Physical examination revealed pain on pelvic compression. We observed an asymmetrical appearance between shoulders; the left shoulder was superior when compared with the right side (Figure 1). Palpation of the scapula aggravated the pain. The inferior angle of the left scapula was not palpable, and the medial border was palpated through the intercostal space between the third and fourth ribs.

Initial radiographs showed additional left olecranon and bilateral ramus pubis fractures. A chest radiograph showed nondisplaced fractures of the second and third ribs without any obvious hemothorax or pneumothorax. No other pathology involving the chest, such as resection of the ribs or congenital anomaly, was observed. The patient reported no history of thoracic trauma or lung surgery. There were no fractures of the scapula, humerus, or clavicles. Thoracic computed tomography was performed, and 3-dimensional (3D) reconstruction showed that the inferior angle of scapula penetrated into the thoracic cavity through the third intercostal space (Figure 2).

Given the intrathoracic scapular dislocation diagnosis, closed reduction under sedation was planned. The patient was placed in the supine position, and reduction was performed by applying pressure on the shoulder anteriorly. This maneuver increased deformity. At the same time, another physician pulled the spine of the scapula superiorly, releasing the scapula out of the thoracic cavity. When the arm was slightly lowered to neutral position, scapular deformity was no longer present (Figure 3). A shoulder sling was applied, and the patient was hospitalized for surgical fixation of pelvic and olecranon fractures. The arm was immobilized in a sling for 1 week, and shoulder exercises were started immediately afterward.

At 1-month follow-up, full shoulder range of motion was achieved, although rehabilitation for the elbow continued. Final follow-up examination at 4 months revealed no difference between shoulders, and no recurrence occurred.

Discussion

Intrathoracic scapular dislocation is a rare injury. There are only a few cases reported in the literature, and most of them are well associated with a predisposing factor. Nettrour and colleagues¹ described the first intrathoracic scapular dislocation, which occurred 6 weeks after sternoclavicular separation and fracture of a rib. In the case reports of Ward and colleagues² and Fowler and colleagues,³ the predisposing factor was resection of the ribs due to pancoast tumor and breast carcinoma, respectively. The mechanism of these dislocations depends on a weak area over the thoracic cage, creating a fulcrum point for levering the scapula into the thoracic cavity.

There are other cases of scapular dislocations that are accompanied by a comminuted fracture of scapula; a review of the literature revealed 3 cases.^4-6 In our opinion, fracture of the inferior pole of the scapula leads to injury of the soft tissues and also results in intrathoracic impaction by creating a weak area over the thoracic cavity. This mechanism can be referred to as penetration.

Our case is singular because it is the first case that is not associated with fracture of the scapula or predisposing factors. Consequently, we report the first pure intrathoracic scapular dislocation in the literature. It is important to suspect intrathoracic scapular dislocation in the case of deformity (Figure 1), even in the absence of any predisposing factors or scapular fracture.

Although plain radiographs may not be elucidative, 3D reconstruction of computed tomography (Figure 2) reveals the pathology and plays an important role in guiding treatment.

In the treatment of our patient, relying on the unique dislocation mechanism without any fracture of the scapula or ribs, we started early active shoulder movement after 1 week of immobilization in a shoulder sling, which prevented recurrence of dislocation. In addition to presenting the first pure intrathoracic scapular dislocation, this case demonstrated satisfactory clinical results with short-term immobilization and early rehabilitation.

Conclusion

Contrary to the literature, the possibility of intrathoracic scapular dislocation should be considered in the differential diagnosis even in patients with a negative anamnesis for predisposing factors, such as lung or chest surgery, and when no fractures are detected. Shoulder or thorax computed tomography, especially 3D reconstructions, are helpful in diagnosing the condition and in guiding treatment. Closed reduction under sedation followed by early rehabilitation is an appropriate treatment method, which resulted in a full return of function in 1 month in our patient.

Scapular dislocation, which is also termed locked scapula or scapulothoracic dislocation, is an unusual condition that can be described as extrathoracic or intrathoracic dislocation, depending on the penetration of scapula into the thoracic cavity.

There have been 3 reported cases of intrathoracic scapular dislocations in the literature,^1-3all associated with a preexisting condition (eg, sternoclavicular separation, prior rib fracture, thoracotomy for a lung transplant procedure, or surgical resection of superior ribs during breast or pulmonary tumor excisions). Three published cases of intrathoracic scapular impaction involve comminuted scapular fractures with intrathoracic impaction of the inferior fragment through intercostal space.^4-6

Here we report an intrathoracic scapular dislocation that was not associated with fracture of the scapula or predisposing factors. To our knowledge, this is the first case of pure intrathoracic dislocation. The possibility of intrathoracic scapular dislocation should be considered as part of the differential diagnosis even in patients with a negative anamnesis for predisposing factors, such as lung or chest surgery. The patient provided written informed consent for print and electronic publication of this case report.

Case Report

A 29-year-old woman presented to the emergency department after a motor vehicle accident. She had tenderness over the left shoulder and left elbow with decreased range of motion; however, motor and sensory examination of the wrist and fingers were normal. No distal neurovascular deficit was noted.

Physical examination revealed pain on pelvic compression. We observed an asymmetrical appearance between shoulders; the left shoulder was superior when compared with the right side (Figure 1). Palpation of the scapula aggravated the pain. The inferior angle of the left scapula was not palpable, and the medial border was palpated through the intercostal space between the third and fourth ribs.

Initial radiographs showed additional left olecranon and bilateral ramus pubis fractures. A chest radiograph showed nondisplaced fractures of the second and third ribs without any obvious hemothorax or pneumothorax. No other pathology involving the chest, such as resection of the ribs or congenital anomaly, was observed. The patient reported no history of thoracic trauma or lung surgery. There were no fractures of the scapula, humerus, or clavicles. Thoracic computed tomography was performed, and 3-dimensional (3D) reconstruction showed that the inferior angle of scapula penetrated into the thoracic cavity through the third intercostal space (Figure 2).

Given the intrathoracic scapular dislocation diagnosis, closed reduction under sedation was planned. The patient was placed in the supine position, and reduction was performed by applying pressure on the shoulder anteriorly. This maneuver increased deformity. At the same time, another physician pulled the spine of the scapula superiorly, releasing the scapula out of the thoracic cavity. When the arm was slightly lowered to neutral position, scapular deformity was no longer present (Figure 3). A shoulder sling was applied, and the patient was hospitalized for surgical fixation of pelvic and olecranon fractures. The arm was immobilized in a sling for 1 week, and shoulder exercises were started immediately afterward.

At 1-month follow-up, full shoulder range of motion was achieved, although rehabilitation for the elbow continued. Final follow-up examination at 4 months revealed no difference between shoulders, and no recurrence occurred.

Discussion

Intrathoracic scapular dislocation is a rare injury. There are only a few cases reported in the literature, and most of them are well associated with a predisposing factor. Nettrour and colleagues¹ described the first intrathoracic scapular dislocation, which occurred 6 weeks after sternoclavicular separation and fracture of a rib. In the case reports of Ward and colleagues² and Fowler and colleagues,³ the predisposing factor was resection of the ribs due to pancoast tumor and breast carcinoma, respectively. The mechanism of these dislocations depends on a weak area over the thoracic cage, creating a fulcrum point for levering the scapula into the thoracic cavity.

There are other cases of scapular dislocations that are accompanied by a comminuted fracture of scapula; a review of the literature revealed 3 cases.^4-6 In our opinion, fracture of the inferior pole of the scapula leads to injury of the soft tissues and also results in intrathoracic impaction by creating a weak area over the thoracic cavity. This mechanism can be referred to as penetration.

Our case is singular because it is the first case that is not associated with fracture of the scapula or predisposing factors. Consequently, we report the first pure intrathoracic scapular dislocation in the literature. It is important to suspect intrathoracic scapular dislocation in the case of deformity (Figure 1), even in the absence of any predisposing factors or scapular fracture.

Although plain radiographs may not be elucidative, 3D reconstruction of computed tomography (Figure 2) reveals the pathology and plays an important role in guiding treatment.

In the treatment of our patient, relying on the unique dislocation mechanism without any fracture of the scapula or ribs, we started early active shoulder movement after 1 week of immobilization in a shoulder sling, which prevented recurrence of dislocation. In addition to presenting the first pure intrathoracic scapular dislocation, this case demonstrated satisfactory clinical results with short-term immobilization and early rehabilitation.

Conclusion

Contrary to the literature, the possibility of intrathoracic scapular dislocation should be considered in the differential diagnosis even in patients with a negative anamnesis for predisposing factors, such as lung or chest surgery, and when no fractures are detected. Shoulder or thorax computed tomography, especially 3D reconstructions, are helpful in diagnosing the condition and in guiding treatment. Closed reduction under sedation followed by early rehabilitation is an appropriate treatment method, which resulted in a full return of function in 1 month in our patient.

Scapular dislocation, which is also termed locked scapula or scapulothoracic dislocation, is an unusual condition that can be described as extrathoracic or intrathoracic dislocation, depending on the penetration of scapula into the thoracic cavity.

There have been 3 reported cases of intrathoracic scapular dislocations in the literature,^1-3all associated with a preexisting condition (eg, sternoclavicular separation, prior rib fracture, thoracotomy for a lung transplant procedure, or surgical resection of superior ribs during breast or pulmonary tumor excisions). Three published cases of intrathoracic scapular impaction involve comminuted scapular fractures with intrathoracic impaction of the inferior fragment through intercostal space.^4-6

Here we report an intrathoracic scapular dislocation that was not associated with fracture of the scapula or predisposing factors. To our knowledge, this is the first case of pure intrathoracic dislocation. The possibility of intrathoracic scapular dislocation should be considered as part of the differential diagnosis even in patients with a negative anamnesis for predisposing factors, such as lung or chest surgery. The patient provided written informed consent for print and electronic publication of this case report.

Case Report

A 29-year-old woman presented to the emergency department after a motor vehicle accident. She had tenderness over the left shoulder and left elbow with decreased range of motion; however, motor and sensory examination of the wrist and fingers were normal. No distal neurovascular deficit was noted.

Physical examination revealed pain on pelvic compression. We observed an asymmetrical appearance between shoulders; the left shoulder was superior when compared with the right side (Figure 1). Palpation of the scapula aggravated the pain. The inferior angle of the left scapula was not palpable, and the medial border was palpated through the intercostal space between the third and fourth ribs.

Initial radiographs showed additional left olecranon and bilateral ramus pubis fractures. A chest radiograph showed nondisplaced fractures of the second and third ribs without any obvious hemothorax or pneumothorax. No other pathology involving the chest, such as resection of the ribs or congenital anomaly, was observed. The patient reported no history of thoracic trauma or lung surgery. There were no fractures of the scapula, humerus, or clavicles. Thoracic computed tomography was performed, and 3-dimensional (3D) reconstruction showed that the inferior angle of scapula penetrated into the thoracic cavity through the third intercostal space (Figure 2).

Given the intrathoracic scapular dislocation diagnosis, closed reduction under sedation was planned. The patient was placed in the supine position, and reduction was performed by applying pressure on the shoulder anteriorly. This maneuver increased deformity. At the same time, another physician pulled the spine of the scapula superiorly, releasing the scapula out of the thoracic cavity. When the arm was slightly lowered to neutral position, scapular deformity was no longer present (Figure 3). A shoulder sling was applied, and the patient was hospitalized for surgical fixation of pelvic and olecranon fractures. The arm was immobilized in a sling for 1 week, and shoulder exercises were started immediately afterward.

At 1-month follow-up, full shoulder range of motion was achieved, although rehabilitation for the elbow continued. Final follow-up examination at 4 months revealed no difference between shoulders, and no recurrence occurred.

Discussion

Intrathoracic scapular dislocation is a rare injury. There are only a few cases reported in the literature, and most of them are well associated with a predisposing factor. Nettrour and colleagues¹ described the first intrathoracic scapular dislocation, which occurred 6 weeks after sternoclavicular separation and fracture of a rib. In the case reports of Ward and colleagues² and Fowler and colleagues,³ the predisposing factor was resection of the ribs due to pancoast tumor and breast carcinoma, respectively. The mechanism of these dislocations depends on a weak area over the thoracic cage, creating a fulcrum point for levering the scapula into the thoracic cavity.

There are other cases of scapular dislocations that are accompanied by a comminuted fracture of scapula; a review of the literature revealed 3 cases.^4-6 In our opinion, fracture of the inferior pole of the scapula leads to injury of the soft tissues and also results in intrathoracic impaction by creating a weak area over the thoracic cavity. This mechanism can be referred to as penetration.

Our case is singular because it is the first case that is not associated with fracture of the scapula or predisposing factors. Consequently, we report the first pure intrathoracic scapular dislocation in the literature. It is important to suspect intrathoracic scapular dislocation in the case of deformity (Figure 1), even in the absence of any predisposing factors or scapular fracture.

Although plain radiographs may not be elucidative, 3D reconstruction of computed tomography (Figure 2) reveals the pathology and plays an important role in guiding treatment.

In the treatment of our patient, relying on the unique dislocation mechanism without any fracture of the scapula or ribs, we started early active shoulder movement after 1 week of immobilization in a shoulder sling, which prevented recurrence of dislocation. In addition to presenting the first pure intrathoracic scapular dislocation, this case demonstrated satisfactory clinical results with short-term immobilization and early rehabilitation.

Conclusion

Contrary to the literature, the possibility of intrathoracic scapular dislocation should be considered in the differential diagnosis even in patients with a negative anamnesis for predisposing factors, such as lung or chest surgery, and when no fractures are detected. Shoulder or thorax computed tomography, especially 3D reconstructions, are helpful in diagnosing the condition and in guiding treatment. Closed reduction under sedation followed by early rehabilitation is an appropriate treatment method, which resulted in a full return of function in 1 month in our patient.

The Internet is becoming a primary source for obtaining medical information. This growing trend may have serious implications for the medical field. As patients increasingly regard the Internet as an essential tool for obtaining health-related information, questions have been raised regarding the quality of medical information available on the Internet.¹ Studies have shown that health-related sites often present inaccurate, inconsistent, and outdated information that may have a negative impact on health care decisions made by patients.²

According to the US Census Bureau, 71.7% of American households report having access to the Internet.³ Of those who have access to Internet, approximately 72% have sought health information online over the last year.⁴ Among people older than age 65 years living in the United States, there has been a growing trend toward using the Internet, from 14% in 2000 to almost 60% in 2013, according to the Pew Research Internet Project.⁵ Most medical websites are viewed for information on diseases and treatment options.⁶ Since most patients want to be informed about treatment options, as well as risks and benefits for each treatment, access to credible information is essential for proper decision-making.⁷

To assess the quality of information on the Internet, we used DISCERN, a standardized questionnaire to aid consumers in judging Internet content.⁸ The DISCERN instrument, available at www.discern.org.uk, was designed by an expert group in the United Kingdom. First, an expert panel developed and tested the instrument, and then health care providers and self-help group members tested it further.^8,9 The questionnaire had been found to have good interrater reliability, regardless of use by health professionals or consumers.^8-10

More than 53,000 shoulder arthroplasties are performed in the United States annually, and the number is growing, with the main goal of pain relief from glenohumeral degenerative joint disease.^11,12 The Internet has become a quasi–second opinion for patients trying to participate in their care. Given the prevalence of shoulder-related surgeries, it is critical to analyze and become familiar with the quality of information that patients read online in order to direct them to nonbiased, all-inclusive websites. In this study, we provide a summary assessment and comparison of the quality of online information pertaining to shoulder replacement, using medical (total shoulder replacement) and nontechnical (shoulder replacement) search terms.

Methods

Websites were identified using 3 search engines (Google, Yahoo, and Bing) and 2 search terms, shoulder replacement (SR) and total shoulder arthroplasty (TSA), on January 17, 2014. These 3 search engines were used because 77% of health care–related information online searches begin through a search engine (Google, Bing, Yahoo); only 13% begin at a health care–specialized website.⁴ These search terms were used after consulting with orthopedic residents and attending physicians in a focus group regarding the terminology used with patients. The first 30 websites in each search engine were identified consecutively and evaluated for category and quality of information using the DISCERN instrument.

A total of 180 websites (90 per search term) were reviewed. Each website was evaluated independently by 3 medical students. In the PRISMA (Preferred Reporting Items for Systematic Reviews and Meta-Analyses) flow diagram, we recorded how websites were identified, screened, and included (Figure 1).13 Websites that were duplicated within each search term and those that were inaccessible were used to determine the total number of noncommercial versus commercial websites, but were excluded from the final analysis. The first part of the analysis involved determining the type of website (eg, commercial vs noncommercial) based upon the html endings. All .com endings were classified as commercial websites; noncommercial included .gov, .org, .edu, and .net endings. Next, each website was categorized based on the target audience. Websites were grouped into health professional–oriented information, patient-oriented, advertisement, or “other.” These classifications were based on those described in previous works.^14,15 The “other” category included images, YouTube videos, another search engine, and open forums, which were also excluded from the final analysis because they were not easily evaluable with the DISCERN instrument. Websites were considered health professional–oriented if they included journal articles, scholarly articles, and/or rehabilitation protocols. Patient-directed websites clearly stated the information was directed to patients or provided a general overview. Advertisement included sites that displayed ads or products for sale. Websites were evaluated for quality using the DISCERN instrument (Figure 2).

DISCERN has 3 subdivision scores: the reliable score (composed of the first 8 questions), the treatment options (the next 7 questions), and 1 final question that addresses the overall quality of the website and is rated independently of the first 15 questions. DISCERN uses 2 scales, a binary scale anchored on both extremes with the number 1 equaling complete absence of the criteria being measured, and the number 5 at the upper extreme, representing completeness of the quality being assessed. In between 1 and 5 is a partial ordinal scale measuring from 2 to 4, which indicates the information is present to some extent but not complete. The ordinal scale allows ranking of the criteria being assessed. Summarizing values from each of the 2 scales poses some concern: the scale is not a true binary scale because of the ordinal scale of the middle numbers (2-4), and as such, is not amenable to being an interval scale to calculate arithmetic means. To summarize the values from the 2 scales, we calculated the harmonic mean, the arithmetic mean, the geometric mean, and the median. The means were empirically compared with the median, and we used the harmonic mean to summarize scale values because it was the best approximation of the medians.

Results

A total of 90 websites were assessed with the search term total shoulder arthroplasty and another 90 with shoulder replacement. When 37 duplicate websites for TSA and 52 for SR were eliminated, 53 (59%) and 38 (42%) unique websites were evaluated for each search term, respectively (Figure 1). (These unique websites are included in the Appendix.) Between the 2 search terms, 20 websites were duplicated. Figure 3 shows the distribution of websites by category. Total shoulder arthroplasty provided the highest percentage of health professional–oriented information; SR had the greatest percentage of patient-oriented information. Both TSA and SR had nearly the same number of advertisements and websites labeled “other.” The percentage of noncommercial websites from each search engine is represented in Figure 4. For SR, Google had 40% (12/30) noncommercial websites compared with Yahoo at 53% (16/30) and Bing at 46% (14/30). Total shoulder arthroplasty had 43% (13/30) noncommercial websites on Google, 27% (8/30) on Yahoo, and 40% (12/30) on Bing. In total, SR had more noncommercial websites, 47% (42/90), compared with 37% (33/90) for TSA.

The mean of all 3 raters for reliablity (DISCERN questions 1-8) and treatment options (DISCERN questions 9-15) is represented in the Table. For both search terms, we found that websites identified as health professional–oriented had the highest reliable mean scores, followed by patient-oriented, and advertisement at the lowest (SR: P = .054; TSA: P = .134). For SR, treatment mean scores demonstrated similar results with health professional–oriented websites receiving the highest, followed by patient-oriented and advertisement (P = .005). However, the treatment mean scores for TSA differed with patient-oriented websites receiving higher scores than health professional–oriented websites, but this was not statistically significant (P= .407). Regarding search terms, there were no significant differences between mean reliable and treatment scores across all categories.

The average overall DISCERN score for TSA websites was 2.5 (range, 1-5), compared with 2.3 (range, 1-5) for SR websites. The overall reliable score (DISCERN questions 1-8) for TSA websites was 2.6 and 2.5 for SR websites (P < .001). For TSA websites, 38% (20/53) were classified as good, having an overall DISCERN score ≥3, versus 26% (10/38) of SR websites. The overall DISCERN score for health professional–oriented websites was 2.7, patient-oriented websites received a score of 2.6, and advertisements had the lowest score at 2.4.

Discussion

Both patients and health professionals obtain information on health care subjects through the Internet, which has become the primary resource for patients.^15,16 However, there are no strict regulations of the content being written. This creates a challenge for the typical user to find credible and evidence-based information, which is important because misleading information could cause undue anxiety, among other effects.^17,18 The aims of this study were to determine the quality of Internet information for shoulder replacement surgeries using the medical terminology total shoulder arthroplasty (TSA) and the nontechnical term shoulder replacement (SR), and to compare the results.

After analyzing the types of websites returned for both total shoulder arthroplasty and shoulder replacement (Figure 4), it was interesting to find that using nonmedical terminology as the search term provided more noncommercial websites compared with total shoulder arthroplasty. Furthermore, Yahoo provided the highest yield of noncommercial websites at 16, with Bing at 14, when using SR as the search term. We believe the increase in noncommercial websites returned for SR was greater than for TSA because SR yielded more patient-oriented websites, which usually had html endings of .edu and .org, as shown in Figure 3 (48% of SR websites offered patient-oriented information).

Although there were more noncommercial websites for SR, the majority of the DISCERN values between the 2 search terms did not differ significantly. This is a direct result of the number of sites (20) that were duplicated across both search terms. However as seen in the Table, TSA had similar reliable mean scores for advertisements and patient-oriented websites but a slightly higher reliable score for health professional–oriented websites. We correlated this with the increased number of health professional–oriented websites returned when using TSA as the search term (Figure 3). The health professional–oriented websites explained their aims and cited their sources more consistently than did patient-oriented sites and advertisements, resulting in higher reliable scores. Although patient-oriented websites frequently lacked citations, they provided information about multiple treatment options, which were more relevant to consumers. This resulted in nearly equivalent reliable scores. Treatment means for advertisements in both SR and TSA were similar. However, treatment means for professional-oriented websites in TSA were lower than those for SR because health professional–oriented websites often were only moderately relevant to consumers, with their focus usually on 1 treatment option or on rehabilitation protocols. Although the DISCERN scores were similar between the search terms, total shoulder arthroplasty provided more websites (20) classified as good—overall DISCERN score, ≥3—than SR did (10). Advertisement websites had similar overall DISCERN scores, which we anticipated because most of the advertisements were duplicated across the search terms.

Using the 2 search terms, academic websites and commercial websites, such as WebMD, consistently received higher reliable and overall DISCERN scores. Advertisement websites, which need to deliver a clear message, frequently scored high on explicitly stating their aims and relevance to consumers, but focused on their products without discussing the benefits of other treatment options. This is significant because Internet search engines, such as Google, offer sponsor links for which organizations pay to appear at the top of the search results. This creates the potential for consumers to receive biased information because most individuals only visit the top 10 websites generated by a search engine.¹⁹

We concluded that the quality of online information relating to SR and TSA was highly variable and frequently of moderate-to-poor quality, with most overall DISCERN scores <3. The quality of information found online for this study using the DISCERN instrument is consistent with those studies using DISCERN to evaluate other medical conditions (eg, bunions, chronic pain, general anesthesia, and anterior cruciate ligament reconstruction).^2,9,15,19 These studies also concluded that online information varies tremendously in quality and completeness.

This study has several limitations. Websites were searched at a single time point and, because Internet resources are frequently updated, the results of this study could vary. Furthermore, although Google, Yahoo, and Bing are 3 of the most popular search engines, these are not the only resources patients use when searching the Internet for health-related information. Other search engines, such as Pubmed.gov and MSN.com, could provide additional websites for Internet users. Lastly, although DISCERN is validated to address the quality of information available online, it does not evaluate the accuracy of the information.⁸ Our use of DISCERN involves 2 scales, a binary yes/no (ratings, 1 and 5) and an ordinal scale (ratings, 2-4). As such, a single mean summary statistic cannot be calculated.

Conclusion

The information available on the Internet pertaining to TSA and SR is highly variable and provides mostly moderate-to-poor quality information based on the DISCERN instrument. Many websites failed to describe the benefits and the risks of different treatment options, including nonoperative management. Health care professionals should be aware that patients often refer to the Internet as a primary resource for obtaining medical information. It is important to direct patients to websites that provide accurate information, because patients who educate themselves about their conditions and actively participate in decision-making may have improved health outcomes.20-22 Overall, academic websites and commercial websites, such as WebMD and OrthoInfo, generally had higher DISCERN scores when using either search term. Of major concern is the potential for misleading advertisements or incorrect information that can negatively affect health outcomes. This study found that using nonmedical terminology (SR) provided more noncommercial and patient-oriented websites, especially through Yahoo. This study highlights the need for more comprehensive online information pertaining to shoulder replacement that can better serve as a resource for Internet users.

The Internet is becoming a primary source for obtaining medical information. This growing trend may have serious implications for the medical field. As patients increasingly regard the Internet as an essential tool for obtaining health-related information, questions have been raised regarding the quality of medical information available on the Internet.¹ Studies have shown that health-related sites often present inaccurate, inconsistent, and outdated information that may have a negative impact on health care decisions made by patients.²

According to the US Census Bureau, 71.7% of American households report having access to the Internet.³ Of those who have access to Internet, approximately 72% have sought health information online over the last year.⁴ Among people older than age 65 years living in the United States, there has been a growing trend toward using the Internet, from 14% in 2000 to almost 60% in 2013, according to the Pew Research Internet Project.⁵ Most medical websites are viewed for information on diseases and treatment options.⁶ Since most patients want to be informed about treatment options, as well as risks and benefits for each treatment, access to credible information is essential for proper decision-making.⁷

To assess the quality of information on the Internet, we used DISCERN, a standardized questionnaire to aid consumers in judging Internet content.⁸ The DISCERN instrument, available at www.discern.org.uk, was designed by an expert group in the United Kingdom. First, an expert panel developed and tested the instrument, and then health care providers and self-help group members tested it further.^8,9 The questionnaire had been found to have good interrater reliability, regardless of use by health professionals or consumers.^8-10

More than 53,000 shoulder arthroplasties are performed in the United States annually, and the number is growing, with the main goal of pain relief from glenohumeral degenerative joint disease.^11,12 The Internet has become a quasi–second opinion for patients trying to participate in their care. Given the prevalence of shoulder-related surgeries, it is critical to analyze and become familiar with the quality of information that patients read online in order to direct them to nonbiased, all-inclusive websites. In this study, we provide a summary assessment and comparison of the quality of online information pertaining to shoulder replacement, using medical (total shoulder replacement) and nontechnical (shoulder replacement) search terms.

Methods

Websites were identified using 3 search engines (Google, Yahoo, and Bing) and 2 search terms, shoulder replacement (SR) and total shoulder arthroplasty (TSA), on January 17, 2014. These 3 search engines were used because 77% of health care–related information online searches begin through a search engine (Google, Bing, Yahoo); only 13% begin at a health care–specialized website.⁴ These search terms were used after consulting with orthopedic residents and attending physicians in a focus group regarding the terminology used with patients. The first 30 websites in each search engine were identified consecutively and evaluated for category and quality of information using the DISCERN instrument.

A total of 180 websites (90 per search term) were reviewed. Each website was evaluated independently by 3 medical students. In the PRISMA (Preferred Reporting Items for Systematic Reviews and Meta-Analyses) flow diagram, we recorded how websites were identified, screened, and included (Figure 1).13 Websites that were duplicated within each search term and those that were inaccessible were used to determine the total number of noncommercial versus commercial websites, but were excluded from the final analysis. The first part of the analysis involved determining the type of website (eg, commercial vs noncommercial) based upon the html endings. All .com endings were classified as commercial websites; noncommercial included .gov, .org, .edu, and .net endings. Next, each website was categorized based on the target audience. Websites were grouped into health professional–oriented information, patient-oriented, advertisement, or “other.” These classifications were based on those described in previous works.^14,15 The “other” category included images, YouTube videos, another search engine, and open forums, which were also excluded from the final analysis because they were not easily evaluable with the DISCERN instrument. Websites were considered health professional–oriented if they included journal articles, scholarly articles, and/or rehabilitation protocols. Patient-directed websites clearly stated the information was directed to patients or provided a general overview. Advertisement included sites that displayed ads or products for sale. Websites were evaluated for quality using the DISCERN instrument (Figure 2).

DISCERN has 3 subdivision scores: the reliable score (composed of the first 8 questions), the treatment options (the next 7 questions), and 1 final question that addresses the overall quality of the website and is rated independently of the first 15 questions. DISCERN uses 2 scales, a binary scale anchored on both extremes with the number 1 equaling complete absence of the criteria being measured, and the number 5 at the upper extreme, representing completeness of the quality being assessed. In between 1 and 5 is a partial ordinal scale measuring from 2 to 4, which indicates the information is present to some extent but not complete. The ordinal scale allows ranking of the criteria being assessed. Summarizing values from each of the 2 scales poses some concern: the scale is not a true binary scale because of the ordinal scale of the middle numbers (2-4), and as such, is not amenable to being an interval scale to calculate arithmetic means. To summarize the values from the 2 scales, we calculated the harmonic mean, the arithmetic mean, the geometric mean, and the median. The means were empirically compared with the median, and we used the harmonic mean to summarize scale values because it was the best approximation of the medians.

Results

A total of 90 websites were assessed with the search term total shoulder arthroplasty and another 90 with shoulder replacement. When 37 duplicate websites for TSA and 52 for SR were eliminated, 53 (59%) and 38 (42%) unique websites were evaluated for each search term, respectively (Figure 1). (These unique websites are included in the Appendix.) Between the 2 search terms, 20 websites were duplicated. Figure 3 shows the distribution of websites by category. Total shoulder arthroplasty provided the highest percentage of health professional–oriented information; SR had the greatest percentage of patient-oriented information. Both TSA and SR had nearly the same number of advertisements and websites labeled “other.” The percentage of noncommercial websites from each search engine is represented in Figure 4. For SR, Google had 40% (12/30) noncommercial websites compared with Yahoo at 53% (16/30) and Bing at 46% (14/30). Total shoulder arthroplasty had 43% (13/30) noncommercial websites on Google, 27% (8/30) on Yahoo, and 40% (12/30) on Bing. In total, SR had more noncommercial websites, 47% (42/90), compared with 37% (33/90) for TSA.

The mean of all 3 raters for reliablity (DISCERN questions 1-8) and treatment options (DISCERN questions 9-15) is represented in the Table. For both search terms, we found that websites identified as health professional–oriented had the highest reliable mean scores, followed by patient-oriented, and advertisement at the lowest (SR: P = .054; TSA: P = .134). For SR, treatment mean scores demonstrated similar results with health professional–oriented websites receiving the highest, followed by patient-oriented and advertisement (P = .005). However, the treatment mean scores for TSA differed with patient-oriented websites receiving higher scores than health professional–oriented websites, but this was not statistically significant (P= .407). Regarding search terms, there were no significant differences between mean reliable and treatment scores across all categories.

The average overall DISCERN score for TSA websites was 2.5 (range, 1-5), compared with 2.3 (range, 1-5) for SR websites. The overall reliable score (DISCERN questions 1-8) for TSA websites was 2.6 and 2.5 for SR websites (P < .001). For TSA websites, 38% (20/53) were classified as good, having an overall DISCERN score ≥3, versus 26% (10/38) of SR websites. The overall DISCERN score for health professional–oriented websites was 2.7, patient-oriented websites received a score of 2.6, and advertisements had the lowest score at 2.4.

Discussion

Both patients and health professionals obtain information on health care subjects through the Internet, which has become the primary resource for patients.^15,16 However, there are no strict regulations of the content being written. This creates a challenge for the typical user to find credible and evidence-based information, which is important because misleading information could cause undue anxiety, among other effects.^17,18 The aims of this study were to determine the quality of Internet information for shoulder replacement surgeries using the medical terminology total shoulder arthroplasty (TSA) and the nontechnical term shoulder replacement (SR), and to compare the results.

After analyzing the types of websites returned for both total shoulder arthroplasty and shoulder replacement (Figure 4), it was interesting to find that using nonmedical terminology as the search term provided more noncommercial websites compared with total shoulder arthroplasty. Furthermore, Yahoo provided the highest yield of noncommercial websites at 16, with Bing at 14, when using SR as the search term. We believe the increase in noncommercial websites returned for SR was greater than for TSA because SR yielded more patient-oriented websites, which usually had html endings of .edu and .org, as shown in Figure 3 (48% of SR websites offered patient-oriented information).

Although there were more noncommercial websites for SR, the majority of the DISCERN values between the 2 search terms did not differ significantly. This is a direct result of the number of sites (20) that were duplicated across both search terms. However as seen in the Table, TSA had similar reliable mean scores for advertisements and patient-oriented websites but a slightly higher reliable score for health professional–oriented websites. We correlated this with the increased number of health professional–oriented websites returned when using TSA as the search term (Figure 3). The health professional–oriented websites explained their aims and cited their sources more consistently than did patient-oriented sites and advertisements, resulting in higher reliable scores. Although patient-oriented websites frequently lacked citations, they provided information about multiple treatment options, which were more relevant to consumers. This resulted in nearly equivalent reliable scores. Treatment means for advertisements in both SR and TSA were similar. However, treatment means for professional-oriented websites in TSA were lower than those for SR because health professional–oriented websites often were only moderately relevant to consumers, with their focus usually on 1 treatment option or on rehabilitation protocols. Although the DISCERN scores were similar between the search terms, total shoulder arthroplasty provided more websites (20) classified as good—overall DISCERN score, ≥3—than SR did (10). Advertisement websites had similar overall DISCERN scores, which we anticipated because most of the advertisements were duplicated across the search terms.

Using the 2 search terms, academic websites and commercial websites, such as WebMD, consistently received higher reliable and overall DISCERN scores. Advertisement websites, which need to deliver a clear message, frequently scored high on explicitly stating their aims and relevance to consumers, but focused on their products without discussing the benefits of other treatment options. This is significant because Internet search engines, such as Google, offer sponsor links for which organizations pay to appear at the top of the search results. This creates the potential for consumers to receive biased information because most individuals only visit the top 10 websites generated by a search engine.¹⁹

We concluded that the quality of online information relating to SR and TSA was highly variable and frequently of moderate-to-poor quality, with most overall DISCERN scores <3. The quality of information found online for this study using the DISCERN instrument is consistent with those studies using DISCERN to evaluate other medical conditions (eg, bunions, chronic pain, general anesthesia, and anterior cruciate ligament reconstruction).^2,9,15,19 These studies also concluded that online information varies tremendously in quality and completeness.

This study has several limitations. Websites were searched at a single time point and, because Internet resources are frequently updated, the results of this study could vary. Furthermore, although Google, Yahoo, and Bing are 3 of the most popular search engines, these are not the only resources patients use when searching the Internet for health-related information. Other search engines, such as Pubmed.gov and MSN.com, could provide additional websites for Internet users. Lastly, although DISCERN is validated to address the quality of information available online, it does not evaluate the accuracy of the information.⁸ Our use of DISCERN involves 2 scales, a binary yes/no (ratings, 1 and 5) and an ordinal scale (ratings, 2-4). As such, a single mean summary statistic cannot be calculated.

Conclusion

The information available on the Internet pertaining to TSA and SR is highly variable and provides mostly moderate-to-poor quality information based on the DISCERN instrument. Many websites failed to describe the benefits and the risks of different treatment options, including nonoperative management. Health care professionals should be aware that patients often refer to the Internet as a primary resource for obtaining medical information. It is important to direct patients to websites that provide accurate information, because patients who educate themselves about their conditions and actively participate in decision-making may have improved health outcomes.20-22 Overall, academic websites and commercial websites, such as WebMD and OrthoInfo, generally had higher DISCERN scores when using either search term. Of major concern is the potential for misleading advertisements or incorrect information that can negatively affect health outcomes. This study found that using nonmedical terminology (SR) provided more noncommercial and patient-oriented websites, especially through Yahoo. This study highlights the need for more comprehensive online information pertaining to shoulder replacement that can better serve as a resource for Internet users.

The Internet is becoming a primary source for obtaining medical information. This growing trend may have serious implications for the medical field. As patients increasingly regard the Internet as an essential tool for obtaining health-related information, questions have been raised regarding the quality of medical information available on the Internet.¹ Studies have shown that health-related sites often present inaccurate, inconsistent, and outdated information that may have a negative impact on health care decisions made by patients.²

According to the US Census Bureau, 71.7% of American households report having access to the Internet.³ Of those who have access to Internet, approximately 72% have sought health information online over the last year.⁴ Among people older than age 65 years living in the United States, there has been a growing trend toward using the Internet, from 14% in 2000 to almost 60% in 2013, according to the Pew Research Internet Project.⁵ Most medical websites are viewed for information on diseases and treatment options.⁶ Since most patients want to be informed about treatment options, as well as risks and benefits for each treatment, access to credible information is essential for proper decision-making.⁷

To assess the quality of information on the Internet, we used DISCERN, a standardized questionnaire to aid consumers in judging Internet content.⁸ The DISCERN instrument, available at www.discern.org.uk, was designed by an expert group in the United Kingdom. First, an expert panel developed and tested the instrument, and then health care providers and self-help group members tested it further.^8,9 The questionnaire had been found to have good interrater reliability, regardless of use by health professionals or consumers.^8-10

More than 53,000 shoulder arthroplasties are performed in the United States annually, and the number is growing, with the main goal of pain relief from glenohumeral degenerative joint disease.^11,12 The Internet has become a quasi–second opinion for patients trying to participate in their care. Given the prevalence of shoulder-related surgeries, it is critical to analyze and become familiar with the quality of information that patients read online in order to direct them to nonbiased, all-inclusive websites. In this study, we provide a summary assessment and comparison of the quality of online information pertaining to shoulder replacement, using medical (total shoulder replacement) and nontechnical (shoulder replacement) search terms.

Methods

Websites were identified using 3 search engines (Google, Yahoo, and Bing) and 2 search terms, shoulder replacement (SR) and total shoulder arthroplasty (TSA), on January 17, 2014. These 3 search engines were used because 77% of health care–related information online searches begin through a search engine (Google, Bing, Yahoo); only 13% begin at a health care–specialized website.⁴ These search terms were used after consulting with orthopedic residents and attending physicians in a focus group regarding the terminology used with patients. The first 30 websites in each search engine were identified consecutively and evaluated for category and quality of information using the DISCERN instrument.

A total of 180 websites (90 per search term) were reviewed. Each website was evaluated independently by 3 medical students. In the PRISMA (Preferred Reporting Items for Systematic Reviews and Meta-Analyses) flow diagram, we recorded how websites were identified, screened, and included (Figure 1).13 Websites that were duplicated within each search term and those that were inaccessible were used to determine the total number of noncommercial versus commercial websites, but were excluded from the final analysis. The first part of the analysis involved determining the type of website (eg, commercial vs noncommercial) based upon the html endings. All .com endings were classified as commercial websites; noncommercial included .gov, .org, .edu, and .net endings. Next, each website was categorized based on the target audience. Websites were grouped into health professional–oriented information, patient-oriented, advertisement, or “other.” These classifications were based on those described in previous works.^14,15 The “other” category included images, YouTube videos, another search engine, and open forums, which were also excluded from the final analysis because they were not easily evaluable with the DISCERN instrument. Websites were considered health professional–oriented if they included journal articles, scholarly articles, and/or rehabilitation protocols. Patient-directed websites clearly stated the information was directed to patients or provided a general overview. Advertisement included sites that displayed ads or products for sale. Websites were evaluated for quality using the DISCERN instrument (Figure 2).

DISCERN has 3 subdivision scores: the reliable score (composed of the first 8 questions), the treatment options (the next 7 questions), and 1 final question that addresses the overall quality of the website and is rated independently of the first 15 questions. DISCERN uses 2 scales, a binary scale anchored on both extremes with the number 1 equaling complete absence of the criteria being measured, and the number 5 at the upper extreme, representing completeness of the quality being assessed. In between 1 and 5 is a partial ordinal scale measuring from 2 to 4, which indicates the information is present to some extent but not complete. The ordinal scale allows ranking of the criteria being assessed. Summarizing values from each of the 2 scales poses some concern: the scale is not a true binary scale because of the ordinal scale of the middle numbers (2-4), and as such, is not amenable to being an interval scale to calculate arithmetic means. To summarize the values from the 2 scales, we calculated the harmonic mean, the arithmetic mean, the geometric mean, and the median. The means were empirically compared with the median, and we used the harmonic mean to summarize scale values because it was the best approximation of the medians.

Results

A total of 90 websites were assessed with the search term total shoulder arthroplasty and another 90 with shoulder replacement. When 37 duplicate websites for TSA and 52 for SR were eliminated, 53 (59%) and 38 (42%) unique websites were evaluated for each search term, respectively (Figure 1). (These unique websites are included in the Appendix.) Between the 2 search terms, 20 websites were duplicated. Figure 3 shows the distribution of websites by category. Total shoulder arthroplasty provided the highest percentage of health professional–oriented information; SR had the greatest percentage of patient-oriented information. Both TSA and SR had nearly the same number of advertisements and websites labeled “other.” The percentage of noncommercial websites from each search engine is represented in Figure 4. For SR, Google had 40% (12/30) noncommercial websites compared with Yahoo at 53% (16/30) and Bing at 46% (14/30). Total shoulder arthroplasty had 43% (13/30) noncommercial websites on Google, 27% (8/30) on Yahoo, and 40% (12/30) on Bing. In total, SR had more noncommercial websites, 47% (42/90), compared with 37% (33/90) for TSA.

The mean of all 3 raters for reliablity (DISCERN questions 1-8) and treatment options (DISCERN questions 9-15) is represented in the Table. For both search terms, we found that websites identified as health professional–oriented had the highest reliable mean scores, followed by patient-oriented, and advertisement at the lowest (SR: P = .054; TSA: P = .134). For SR, treatment mean scores demonstrated similar results with health professional–oriented websites receiving the highest, followed by patient-oriented and advertisement (P = .005). However, the treatment mean scores for TSA differed with patient-oriented websites receiving higher scores than health professional–oriented websites, but this was not statistically significant (P= .407). Regarding search terms, there were no significant differences between mean reliable and treatment scores across all categories.

The average overall DISCERN score for TSA websites was 2.5 (range, 1-5), compared with 2.3 (range, 1-5) for SR websites. The overall reliable score (DISCERN questions 1-8) for TSA websites was 2.6 and 2.5 for SR websites (P < .001). For TSA websites, 38% (20/53) were classified as good, having an overall DISCERN score ≥3, versus 26% (10/38) of SR websites. The overall DISCERN score for health professional–oriented websites was 2.7, patient-oriented websites received a score of 2.6, and advertisements had the lowest score at 2.4.

Discussion

Both patients and health professionals obtain information on health care subjects through the Internet, which has become the primary resource for patients.^15,16 However, there are no strict regulations of the content being written. This creates a challenge for the typical user to find credible and evidence-based information, which is important because misleading information could cause undue anxiety, among other effects.^17,18 The aims of this study were to determine the quality of Internet information for shoulder replacement surgeries using the medical terminology total shoulder arthroplasty (TSA) and the nontechnical term shoulder replacement (SR), and to compare the results.

After analyzing the types of websites returned for both total shoulder arthroplasty and shoulder replacement (Figure 4), it was interesting to find that using nonmedical terminology as the search term provided more noncommercial websites compared with total shoulder arthroplasty. Furthermore, Yahoo provided the highest yield of noncommercial websites at 16, with Bing at 14, when using SR as the search term. We believe the increase in noncommercial websites returned for SR was greater than for TSA because SR yielded more patient-oriented websites, which usually had html endings of .edu and .org, as shown in Figure 3 (48% of SR websites offered patient-oriented information).

Although there were more noncommercial websites for SR, the majority of the DISCERN values between the 2 search terms did not differ significantly. This is a direct result of the number of sites (20) that were duplicated across both search terms. However as seen in the Table, TSA had similar reliable mean scores for advertisements and patient-oriented websites but a slightly higher reliable score for health professional–oriented websites. We correlated this with the increased number of health professional–oriented websites returned when using TSA as the search term (Figure 3). The health professional–oriented websites explained their aims and cited their sources more consistently than did patient-oriented sites and advertisements, resulting in higher reliable scores. Although patient-oriented websites frequently lacked citations, they provided information about multiple treatment options, which were more relevant to consumers. This resulted in nearly equivalent reliable scores. Treatment means for advertisements in both SR and TSA were similar. However, treatment means for professional-oriented websites in TSA were lower than those for SR because health professional–oriented websites often were only moderately relevant to consumers, with their focus usually on 1 treatment option or on rehabilitation protocols. Although the DISCERN scores were similar between the search terms, total shoulder arthroplasty provided more websites (20) classified as good—overall DISCERN score, ≥3—than SR did (10). Advertisement websites had similar overall DISCERN scores, which we anticipated because most of the advertisements were duplicated across the search terms.

Using the 2 search terms, academic websites and commercial websites, such as WebMD, consistently received higher reliable and overall DISCERN scores. Advertisement websites, which need to deliver a clear message, frequently scored high on explicitly stating their aims and relevance to consumers, but focused on their products without discussing the benefits of other treatment options. This is significant because Internet search engines, such as Google, offer sponsor links for which organizations pay to appear at the top of the search results. This creates the potential for consumers to receive biased information because most individuals only visit the top 10 websites generated by a search engine.¹⁹

We concluded that the quality of online information relating to SR and TSA was highly variable and frequently of moderate-to-poor quality, with most overall DISCERN scores <3. The quality of information found online for this study using the DISCERN instrument is consistent with those studies using DISCERN to evaluate other medical conditions (eg, bunions, chronic pain, general anesthesia, and anterior cruciate ligament reconstruction).^2,9,15,19 These studies also concluded that online information varies tremendously in quality and completeness.

This study has several limitations. Websites were searched at a single time point and, because Internet resources are frequently updated, the results of this study could vary. Furthermore, although Google, Yahoo, and Bing are 3 of the most popular search engines, these are not the only resources patients use when searching the Internet for health-related information. Other search engines, such as Pubmed.gov and MSN.com, could provide additional websites for Internet users. Lastly, although DISCERN is validated to address the quality of information available online, it does not evaluate the accuracy of the information.⁸ Our use of DISCERN involves 2 scales, a binary yes/no (ratings, 1 and 5) and an ordinal scale (ratings, 2-4). As such, a single mean summary statistic cannot be calculated.

Conclusion

The information available on the Internet pertaining to TSA and SR is highly variable and provides mostly moderate-to-poor quality information based on the DISCERN instrument. Many websites failed to describe the benefits and the risks of different treatment options, including nonoperative management. Health care professionals should be aware that patients often refer to the Internet as a primary resource for obtaining medical information. It is important to direct patients to websites that provide accurate information, because patients who educate themselves about their conditions and actively participate in decision-making may have improved health outcomes.20-22 Overall, academic websites and commercial websites, such as WebMD and OrthoInfo, generally had higher DISCERN scores when using either search term. Of major concern is the potential for misleading advertisements or incorrect information that can negatively affect health outcomes. This study found that using nonmedical terminology (SR) provided more noncommercial and patient-oriented websites, especially through Yahoo. This study highlights the need for more comprehensive online information pertaining to shoulder replacement that can better serve as a resource for Internet users.

Degenerative arthritis is a widespread chronic condition with an incidence of almost 43 million and annual health care costs of $60 billion in the United States alone.1 Although many cases can be managed symptomatically with medical therapy and intra-articular injections,2 many patients experience disease progression resulting in decreased ambulatory ability and work productivity. For these patients, elective hip and knee arthroplasties can drastically improve quality of life and functionality.3,4 Over the past decade, there has been a marked increase in the number of primary and revision total hip and knee arthroplasties performed in the United States. By 2030, the demand for primary total hip arthroplasties will grow an estimated 174%, to 572,000 procedures. Likewise, the demand for primary total knee arthroplasties is projected to grow by 673%, to 3.48 million procedures.5 However, though better surgical techniques and technology have led to improved functional outcomes, there is still substantial risk for complications in the perioperative period, especially in the geriatric population, in which substantial comorbidities are common.6-9

Acute kidney injury (AKI) is a common public health problem in hospitalized patients and in patients undergoing procedures. More than one-third of all AKI cases occur in surgical settings.^10,11 Over the past decade, both community-acquired and in-hospital AKIs rapidly increased in incidence in all major clinical settings.^12-14 Patients with AKI have high rates of adverse outcomes during hospitalization and discharge.^11,15 Sequelae of AKIs include worsening chronic kidney disease (CKD) and progression to end-stage renal disease, necessitating either long-term dialysis or transplantation.¹² This in turn leads to exacerbated disability, diminished quality of life, and disproportionate burden on health care resources.

Much of our knowledge about postoperative AKI has been derived from cardiovascular, thoracic, and abdominal surgery settings. However, there is a paucity of data on epidemiology and trends for either AKI or associated outcomes in patients undergoing major orthopedic surgery. The few studies to date either were single-center or had inadequate sample sizes for appropriately powered analysis of the risk factors and outcomes related to AKI.¹⁶

In the study reported here, we analyzed a large cohort of patients from a nationwide multicenter database to determine the incidence of and risk factors for AKI. We also examined the mortality and adverse discharges associated with AKI after major joint surgery. Lastly, we assessed temporal trends in both incidence and outcomes of AKI, including the death risk attributable to AKI.

Methods

Database

We extracted our study cohort from the Nationwide Inpatient Sample (NIS) and the National Inpatient Sample of Healthcare Cost and Utilization Project (HCUP) compiled by the Agency for Healthcare Research and Quality.¹⁷ NIS, the largest inpatient care database in the United States, stores data from almost 8 million stays in about 1000 hospitals across the country each year. Its participating hospital pool consists of about 20% of US community hospitals, resulting in a sampling frame comprising about 90% of all hospital discharges in the United States. This allows for calculation of precise, weighted nationwide estimates. Data elements within NIS are drawn from hospital discharge abstracts that indicate all procedures performed. NIS also stores information on patient characteristics, length of stay (LOS), discharge disposition, postoperative morbidity, and observed in-hospital mortality. However, it stores no information on long-term follow-up or complications after discharge.

Data Analysis

For the period 2002–2012, we queried the NIS database for hip and knee arthroplasties with primary diagnosis codes for osteoarthritis and secondary codes for AKI. We excluded patients under age 18 years and patients with diagnosis codes for hip and knee fracture/necrosis, inflammatory/infectious arthritis, or bone neoplasms (Table 1). We then extracted baseline characteristics of the study population. Patient-level characteristics included age, sex, race, quartile classification of median household income according to postal (ZIP) code, and primary payer (Medicare/Medicaid, private insurance, self-pay, no charge). Hospital-level characteristics included hospital location (urban, rural), hospital bed size (small, medium, large), region (Northeast, Midwest/North Central, South, West), and teaching status. We defined illness severity and likelihood of death using Deyo’s modification of the Charlson Comorbidity Index (CCI), which draws on principal and secondary ICD-9-CM (International Classification of Diseases, Ninth Revision-Clinical Modification) diagnosis codes, procedure codes, and patient demographics to estimate a patient’s mortality risk. This method reliably predicts mortality and readmission in the orthopedic population.^18,19 We assessed the effect of AKI on 4 outcomes, including in-hospital mortality, discharge disposition, LOS, and cost of stay. Discharge disposition was grouped by either (a) home or short-term facility or (b) adverse discharge. Home or short-term facility covered routine, short-term hospital, against medical advice, home intravenous provider, another rehabilitation facility, another institution for outpatient services, institution for outpatient services, discharged alive, and destination unknown; adverse discharge covered skilled nursing facility, intermediate care, hospice home, hospice medical facility, long-term care hospital, and certified nursing facility. This dichotomization of discharge disposition is often used in studies of NIS data.²⁰

Statistical Analyses

We compared the baseline characteristics of hospitalized patients with and without AKI. To test for significance, we used the χ² test for categorical variables, the Student t test for normally distributed continuous variables, the Wilcoxon rank sum test for non-normally distributed continuous variables, and the Cochran-Armitage test for trends in AKI incidence. We used survey logistic regression models to calculate adjusted odds ratios (ORs) with 95% confidence intervals (95% CIs) in order to estimate the predictors of AKI and the impact of AKI on hospital outcomes. We constructed final models after adjusting for confounders, testing for potential interactions, and ensuring no multicolinearity between covariates. Last, we computed the risk proportion of death attributable to AKI, indicating the proportion of deaths that could potentially be avoided if AKI and its complications were abrogated.²¹

We performed all statistical analyses with SAS Version 9.3 (SAS Institute) using designated weight values to produce weighted national estimates. The threshold for statistical significance was set at P < .01 (with ORs and 95% CIs that excluded 1).

Results

AKI Incidence, Risk Factors, and Trends

We identified 7,235,251 patients who underwent elective hip or knee arthroplasty for osteoarthritis between 2002 and 2012—an estimate consistent with data from the Centers for Disease Control and Prevention.²² Of that total, 94,367 (1.3%) had AKI. The proportion of discharges diagnosed with AKI increased rapidly over the decade, from 0.5% in 2002 to 1.8% to 1.9% in the period 2010–2012. This upward trend was highly significant (P_trend < .001) (Figure 1). Patients with AKI (vs patients without AKI) were more likely to be older (mean age, 70 vs 66 years; P < .001), male (50.8% vs 38.4%; P < .001), and black (10.07% vs 5.15%; P<. 001). They were also found to have a significantly higher comorbidity score (mean CCI, 2.8 vs 1.5; P < .001) and higher proportions of comorbidities, including hypertension, CKD, atrial fibrillation, diabetes mellitus (DM), congestive heart failure, chronic liver disease, and hepatitis C virus infection. In addition, AKI was associated with perioperative myocardial infarction (MI), sepsis, cardiac catheterization, and blood transfusion. Regarding socioeconomic characteristics, patients with AKI were more likely to have Medicare/Medicaid insurance (72.26% vs 58.06%; P < .001) and to belong to the extremes of income categories (Table 2).

Using multivariable logistic regression, we found that increased age (1.11 increase in adjusted OR for every year older; 95% CI, 1.09-1.14; P < .001), male sex (adjusted OR, 1.65; 95% CI, 1.60-1.71; P < .001), and black race (adjusted OR, 1.57; 95% CI, 1.45-1.69; P < .001) were significantly associated with postoperative AKI. Regarding comorbidities, baseline CKD (adjusted OR, 8.64; 95% CI, 8.14-9.18; P < .001) and congestive heart failure (adjusted OR, 2.74; 95% CI, 2.57-2.92; P< .0001) were most significantly associated with AKI. Perioperative events, including sepsis (adjusted OR, 35.64; 95% CI, 30.28-41.96; P < .0001), MI (adjusted OR, 6.14; 95% CI, 5.17-7.28; P < .0001), and blood transfusion (adjusted OR, 2.28; 95% CI, 2.15-2.42; P < .0001), were also strongly associated with postoperative AKI. Last, compared with urban hospitals and small hospital bed size, rural hospitals (adjusted OR, 0.70; 95% CI, 0.60-0.81; P< .001) and large bed size (adjusted OR, 0.82; 95% CI, 0.70-0.93; P = .003) were associated with lower probability of developing AKI (Table 3).

Figure 2 elucidates the frequency of AKI based on a combination of key preoperative comorbid conditions and postoperative complications—demonstrating that the proportion of AKI cases associated with other postoperative complications is significantly higher in the CKD and concomitant DM/CKD patient populations. Patients hospitalized with CKD exhibited higher rates of AKI in cases involving blood transfusion (20.9% vs 1.8%; P < .001), acute MI (48.9% vs 13.8%; P < .001), and sepsis (74.7% vs 36.3%;P< .001) relative to patients without CKD. Similarly, patients with concomitant DM/CKD exhibited higher rates of AKI in cases involving blood transfusion (23% vs 1.9%; P< .001), acute MI (51.1% vs 12.1%; P< .001), and sepsis (75% vs 38.2%; P < .001) relative to patients without either condition. However, patients hospitalized with DM alone exhibited only marginally higher rates of AKI in cases involving blood transfusion (4.7% vs 2%; P < .01) and acute MI (19.2% vs 16.7%; P< .01) and a lower rate in cases involving sepsis (38.2% vs 41.7%; P < .01) relative to patients without DM. These data suggest that CKD is the most significant clinically relevant risk factor for AKI and that CKD may synergize with DM to raise the risk for AKI.

Outcomes

We then analyzed the impact of AKI on hospital outcomes, including in-hospital mortality, discharge disposition, LOS, and cost of care. Mortality was significantly higher in patients with AKI than in patients without it (2.08% vs 0.06%; P < .001). Even after adjusting for confounders (eg, demographics, comorbidity burden, perioperative sepsis, hospital-level characteristics), AKI was still associated with strikingly higher odds of in-hospital death (adjusted OR, 11.32; 95% CI, 9.34-13.74; P < .001). However, analysis of temporal trends indicated that the odds for adjusted mortality associated with AKI decreased from 18.09 to 9.45 (P_trend = .01) over the period 2002–2012 (Figure 3). This decrease in odds of death was countered by an increase in incidence of AKI, resulting in a stable attributable risk proportion (97.9% in 2002 to 97.3% in 2012; P_trend = .90) (Table 4). Regarding discharge disposition, patients with AKI were much less likely to be discharged home (41.35% vs 62.59%; P < .001) and more likely to be discharged to long-term care (56.37% vs 37.03%; P< .001). After adjustment for confounders, AKI was associated with significantly increased odds of adverse discharge (adjusted OR, 2.24; 95% CI, 2.12-2.36; P< .001). Analysis of temporal trends revealed no appreciable decrease in the adjusted odds of adverse discharge between 2002 (adjusted OR, 1.87; 95% CI, 1.37-2.55; P < .001) and 2012 (adjusted OR, 1.93; 95% CI, 1.76-2.11; P < .001) (Figure 4, Table 5). Last, both mean LOS (5 days vs 3 days; P < .001) and mean cost of hospitalization (US $22,269 vs $15,757; P < .001) were significantly higher in patients with AKI.

Discussion

In this study, we found that the incidence of AKI among hospitalized patients increased 4-fold between 2002 and 2012. Moreover, we identified numerous patient-specific, hospital-specific, perioperative risk factors for AKI. Most important, we found that AKI was associated with a strikingly higher risk of in-hospital death, and surviving patients were more likely to experience adverse discharge. Although the adjusted mortality rate associated with AKI decreased over that decade, the attributable risk proportion remained stable.

Few studies have addressed this significant public health concern. In one recent study in Australia, Kimmel and colleagues¹⁶ identified risk factors for AKI but lacked data on AKI outcomes. In a study of complications and mortality occurring after orthopedic surgery, Belmont and colleagues²² categorized complications as either local or systemic but did not examine renal complications. Only 2 other major studies have been conducted on renal outcomes associated with major joint surgery, and both were limited to patients with acute hip fractures. The first included acute fracture surgery patients and omitted elective joint surgery patients, and it evaluated admission renal function but not postoperative AKI.²² The second study had a sample size of only 170 patients.²³ Thus, the literature leaves us with a crucial knowledge gap in renal outcomes and their postoperative impact in elective arthroplasties.

The present study filled this information gap by examining the incidence, risk factors, outcomes, and temporal trends of AKI after elective hip and knee arthroplasties. The increasing incidence of AKI in this surgical setting is similar to that of AKI in other surgical settings (cardiac and noncardiac).²¹ Although our analysis was limited by lack of perioperative management data, patients undergoing elective joint arthroplasty can experience kidney dysfunction for several reasons, including volume depletion, postoperative sepsis, and influence of medications, such as nonsteroidal anti-inflammatory drugs (NSAIDs), especially in older patients with more comorbidities and a higher burden of CKD. Each of these factors can cause renal dysfunction in patients having orthopedic procedures.²⁴ Moreover, NSAID use among elective joint arthroplasty patients is likely higher because of an emphasis on multimodal analgesia, as recent randomized controlled trials have demonstrated the efficacy of NSAID use in controlling pain without increasing bleeding.^25-27 Our results also demonstrated that the absolute incidence of AKI after orthopedic surgery is relatively low. One possible explanation for this phenomenon is that the definitions used were based on ICD-9-CM codes that underestimate the true incidence of AKI.

Consistent with other studies, we found that certain key preoperative comorbid conditions and postoperative events were associated with higher AKI risk. We stratified the rate of AKI associated with each postoperative event (sepsis, acute MI, cardiac catheterization, need for transfusion) by DM/CKD comorbidity. CKD was associated with significantly higher AKI risk across all postoperative complications. This information may provide clinicians with bedside information that can be used to determine which patients may be at higher or lower risk for AKI.

Our analysis of patient outcomes revealed that, though AKI was relatively uncommon, it increased the risk for death during hospitalization more than 10-fold between 2002 and 2012. Although the adjusted OR of in-hospital mortality decreased over the decade studied, the concurrent increase in AKI incidence caused the attributable risk of death associated with AKI to essentially remain the same. This observation is consistent with recent reports from cardiac surgery settings.²¹ These data together suggest that ameliorating occurrences of AKI would decrease mortality and increase quality of care for patients undergoing elective joint surgeries.

We also examined the effect of AKI on resource use by studying LOS, costs, and risk for adverse discharge. Much as in other surgical settings, AKI increased both LOS and overall hospitalization costs. More important, AKI was associated with increased adverse discharge (discharge to long-term care or nursing homes). Although exact reasons are unclear, we can speculate that postoperative renal dysfunction precludes early rehabilitation, impeding desired functional outcome and disposition.^28,29 Given the projected increases in primary and revision hip and knee arthroplasties,⁵ these data predict that the impact of AKI on health outcomes will increase alarmingly in coming years.

There are limitations to our study. First, it was based on administrative data and lacked patient-level and laboratory data. As reported, the sensitivity of AKI codes remains moderate,³⁰ so the true burden may be higher than indicated here. As the definition of AKI was based on administrative coding, we also could not estimate severity, though previous studies have found that administrative codes typically capture a more severe form of disease.³¹ Another limitation is that, because the data were deidentified, we could not delineate the risk for recurrent AKI in repeated surgical procedures, though this cohort unlikely was large enough to qualitatively affect our results. The third limitation is that, though we used CCI to adjust for the comorbidity burden, we were unable to account for other unmeasured confounders associated with increased AKI incidence, such as specific medication use. In addition, given the lack of patient-level data, we could not analyze the specific factors responsible for AKI in the perioperative period. Nevertheless, the strengths of a nationally representative sample, such as large sample size and generalizability, outweigh these limitations.

Conclusion

AKI is potentially an important quality indicator of elective joint surgery, and reducing its incidence is therefore essential for quality improvement. Given that hip and knee arthroplasties are projected to increase exponentially, as is the burden of comorbid conditions in this population, postoperative AKI will continue to have an incremental impact on health and health care resources. Thus, a carefully planned approach of interdisciplinary perioperative care is warranted to reduce both the risk and the consequences of this devastating condition.

Degenerative arthritis is a widespread chronic condition with an incidence of almost 43 million and annual health care costs of $60 billion in the United States alone.1 Although many cases can be managed symptomatically with medical therapy and intra-articular injections,2 many patients experience disease progression resulting in decreased ambulatory ability and work productivity. For these patients, elective hip and knee arthroplasties can drastically improve quality of life and functionality.3,4 Over the past decade, there has been a marked increase in the number of primary and revision total hip and knee arthroplasties performed in the United States. By 2030, the demand for primary total hip arthroplasties will grow an estimated 174%, to 572,000 procedures. Likewise, the demand for primary total knee arthroplasties is projected to grow by 673%, to 3.48 million procedures.5 However, though better surgical techniques and technology have led to improved functional outcomes, there is still substantial risk for complications in the perioperative period, especially in the geriatric population, in which substantial comorbidities are common.6-9

Acute kidney injury (AKI) is a common public health problem in hospitalized patients and in patients undergoing procedures. More than one-third of all AKI cases occur in surgical settings.^10,11 Over the past decade, both community-acquired and in-hospital AKIs rapidly increased in incidence in all major clinical settings.^12-14 Patients with AKI have high rates of adverse outcomes during hospitalization and discharge.^11,15 Sequelae of AKIs include worsening chronic kidney disease (CKD) and progression to end-stage renal disease, necessitating either long-term dialysis or transplantation.¹² This in turn leads to exacerbated disability, diminished quality of life, and disproportionate burden on health care resources.

Much of our knowledge about postoperative AKI has been derived from cardiovascular, thoracic, and abdominal surgery settings. However, there is a paucity of data on epidemiology and trends for either AKI or associated outcomes in patients undergoing major orthopedic surgery. The few studies to date either were single-center or had inadequate sample sizes for appropriately powered analysis of the risk factors and outcomes related to AKI.¹⁶

In the study reported here, we analyzed a large cohort of patients from a nationwide multicenter database to determine the incidence of and risk factors for AKI. We also examined the mortality and adverse discharges associated with AKI after major joint surgery. Lastly, we assessed temporal trends in both incidence and outcomes of AKI, including the death risk attributable to AKI.

Methods

Database

We extracted our study cohort from the Nationwide Inpatient Sample (NIS) and the National Inpatient Sample of Healthcare Cost and Utilization Project (HCUP) compiled by the Agency for Healthcare Research and Quality.¹⁷ NIS, the largest inpatient care database in the United States, stores data from almost 8 million stays in about 1000 hospitals across the country each year. Its participating hospital pool consists of about 20% of US community hospitals, resulting in a sampling frame comprising about 90% of all hospital discharges in the United States. This allows for calculation of precise, weighted nationwide estimates. Data elements within NIS are drawn from hospital discharge abstracts that indicate all procedures performed. NIS also stores information on patient characteristics, length of stay (LOS), discharge disposition, postoperative morbidity, and observed in-hospital mortality. However, it stores no information on long-term follow-up or complications after discharge.

Data Analysis

For the period 2002–2012, we queried the NIS database for hip and knee arthroplasties with primary diagnosis codes for osteoarthritis and secondary codes for AKI. We excluded patients under age 18 years and patients with diagnosis codes for hip and knee fracture/necrosis, inflammatory/infectious arthritis, or bone neoplasms (Table 1). We then extracted baseline characteristics of the study population. Patient-level characteristics included age, sex, race, quartile classification of median household income according to postal (ZIP) code, and primary payer (Medicare/Medicaid, private insurance, self-pay, no charge). Hospital-level characteristics included hospital location (urban, rural), hospital bed size (small, medium, large), region (Northeast, Midwest/North Central, South, West), and teaching status. We defined illness severity and likelihood of death using Deyo’s modification of the Charlson Comorbidity Index (CCI), which draws on principal and secondary ICD-9-CM (International Classification of Diseases, Ninth Revision-Clinical Modification) diagnosis codes, procedure codes, and patient demographics to estimate a patient’s mortality risk. This method reliably predicts mortality and readmission in the orthopedic population.^18,19 We assessed the effect of AKI on 4 outcomes, including in-hospital mortality, discharge disposition, LOS, and cost of stay. Discharge disposition was grouped by either (a) home or short-term facility or (b) adverse discharge. Home or short-term facility covered routine, short-term hospital, against medical advice, home intravenous provider, another rehabilitation facility, another institution for outpatient services, institution for outpatient services, discharged alive, and destination unknown; adverse discharge covered skilled nursing facility, intermediate care, hospice home, hospice medical facility, long-term care hospital, and certified nursing facility. This dichotomization of discharge disposition is often used in studies of NIS data.²⁰

Statistical Analyses

We compared the baseline characteristics of hospitalized patients with and without AKI. To test for significance, we used the χ² test for categorical variables, the Student t test for normally distributed continuous variables, the Wilcoxon rank sum test for non-normally distributed continuous variables, and the Cochran-Armitage test for trends in AKI incidence. We used survey logistic regression models to calculate adjusted odds ratios (ORs) with 95% confidence intervals (95% CIs) in order to estimate the predictors of AKI and the impact of AKI on hospital outcomes. We constructed final models after adjusting for confounders, testing for potential interactions, and ensuring no multicolinearity between covariates. Last, we computed the risk proportion of death attributable to AKI, indicating the proportion of deaths that could potentially be avoided if AKI and its complications were abrogated.²¹

We performed all statistical analyses with SAS Version 9.3 (SAS Institute) using designated weight values to produce weighted national estimates. The threshold for statistical significance was set at P < .01 (with ORs and 95% CIs that excluded 1).

Results

AKI Incidence, Risk Factors, and Trends

We identified 7,235,251 patients who underwent elective hip or knee arthroplasty for osteoarthritis between 2002 and 2012—an estimate consistent with data from the Centers for Disease Control and Prevention.²² Of that total, 94,367 (1.3%) had AKI. The proportion of discharges diagnosed with AKI increased rapidly over the decade, from 0.5% in 2002 to 1.8% to 1.9% in the period 2010–2012. This upward trend was highly significant (P_trend < .001) (Figure 1). Patients with AKI (vs patients without AKI) were more likely to be older (mean age, 70 vs 66 years; P < .001), male (50.8% vs 38.4%; P < .001), and black (10.07% vs 5.15%; P<. 001). They were also found to have a significantly higher comorbidity score (mean CCI, 2.8 vs 1.5; P < .001) and higher proportions of comorbidities, including hypertension, CKD, atrial fibrillation, diabetes mellitus (DM), congestive heart failure, chronic liver disease, and hepatitis C virus infection. In addition, AKI was associated with perioperative myocardial infarction (MI), sepsis, cardiac catheterization, and blood transfusion. Regarding socioeconomic characteristics, patients with AKI were more likely to have Medicare/Medicaid insurance (72.26% vs 58.06%; P < .001) and to belong to the extremes of income categories (Table 2).

Using multivariable logistic regression, we found that increased age (1.11 increase in adjusted OR for every year older; 95% CI, 1.09-1.14; P < .001), male sex (adjusted OR, 1.65; 95% CI, 1.60-1.71; P < .001), and black race (adjusted OR, 1.57; 95% CI, 1.45-1.69; P < .001) were significantly associated with postoperative AKI. Regarding comorbidities, baseline CKD (adjusted OR, 8.64; 95% CI, 8.14-9.18; P < .001) and congestive heart failure (adjusted OR, 2.74; 95% CI, 2.57-2.92; P< .0001) were most significantly associated with AKI. Perioperative events, including sepsis (adjusted OR, 35.64; 95% CI, 30.28-41.96; P < .0001), MI (adjusted OR, 6.14; 95% CI, 5.17-7.28; P < .0001), and blood transfusion (adjusted OR, 2.28; 95% CI, 2.15-2.42; P < .0001), were also strongly associated with postoperative AKI. Last, compared with urban hospitals and small hospital bed size, rural hospitals (adjusted OR, 0.70; 95% CI, 0.60-0.81; P< .001) and large bed size (adjusted OR, 0.82; 95% CI, 0.70-0.93; P = .003) were associated with lower probability of developing AKI (Table 3).

Figure 2 elucidates the frequency of AKI based on a combination of key preoperative comorbid conditions and postoperative complications—demonstrating that the proportion of AKI cases associated with other postoperative complications is significantly higher in the CKD and concomitant DM/CKD patient populations. Patients hospitalized with CKD exhibited higher rates of AKI in cases involving blood transfusion (20.9% vs 1.8%; P < .001), acute MI (48.9% vs 13.8%; P < .001), and sepsis (74.7% vs 36.3%;P< .001) relative to patients without CKD. Similarly, patients with concomitant DM/CKD exhibited higher rates of AKI in cases involving blood transfusion (23% vs 1.9%; P< .001), acute MI (51.1% vs 12.1%; P< .001), and sepsis (75% vs 38.2%; P < .001) relative to patients without either condition. However, patients hospitalized with DM alone exhibited only marginally higher rates of AKI in cases involving blood transfusion (4.7% vs 2%; P < .01) and acute MI (19.2% vs 16.7%; P< .01) and a lower rate in cases involving sepsis (38.2% vs 41.7%; P < .01) relative to patients without DM. These data suggest that CKD is the most significant clinically relevant risk factor for AKI and that CKD may synergize with DM to raise the risk for AKI.

Outcomes

We then analyzed the impact of AKI on hospital outcomes, including in-hospital mortality, discharge disposition, LOS, and cost of care. Mortality was significantly higher in patients with AKI than in patients without it (2.08% vs 0.06%; P < .001). Even after adjusting for confounders (eg, demographics, comorbidity burden, perioperative sepsis, hospital-level characteristics), AKI was still associated with strikingly higher odds of in-hospital death (adjusted OR, 11.32; 95% CI, 9.34-13.74; P < .001). However, analysis of temporal trends indicated that the odds for adjusted mortality associated with AKI decreased from 18.09 to 9.45 (P_trend = .01) over the period 2002–2012 (Figure 3). This decrease in odds of death was countered by an increase in incidence of AKI, resulting in a stable attributable risk proportion (97.9% in 2002 to 97.3% in 2012; P_trend = .90) (Table 4). Regarding discharge disposition, patients with AKI were much less likely to be discharged home (41.35% vs 62.59%; P < .001) and more likely to be discharged to long-term care (56.37% vs 37.03%; P< .001). After adjustment for confounders, AKI was associated with significantly increased odds of adverse discharge (adjusted OR, 2.24; 95% CI, 2.12-2.36; P< .001). Analysis of temporal trends revealed no appreciable decrease in the adjusted odds of adverse discharge between 2002 (adjusted OR, 1.87; 95% CI, 1.37-2.55; P < .001) and 2012 (adjusted OR, 1.93; 95% CI, 1.76-2.11; P < .001) (Figure 4, Table 5). Last, both mean LOS (5 days vs 3 days; P < .001) and mean cost of hospitalization (US $22,269 vs $15,757; P < .001) were significantly higher in patients with AKI.

Discussion

In this study, we found that the incidence of AKI among hospitalized patients increased 4-fold between 2002 and 2012. Moreover, we identified numerous patient-specific, hospital-specific, perioperative risk factors for AKI. Most important, we found that AKI was associated with a strikingly higher risk of in-hospital death, and surviving patients were more likely to experience adverse discharge. Although the adjusted mortality rate associated with AKI decreased over that decade, the attributable risk proportion remained stable.

Few studies have addressed this significant public health concern. In one recent study in Australia, Kimmel and colleagues¹⁶ identified risk factors for AKI but lacked data on AKI outcomes. In a study of complications and mortality occurring after orthopedic surgery, Belmont and colleagues²² categorized complications as either local or systemic but did not examine renal complications. Only 2 other major studies have been conducted on renal outcomes associated with major joint surgery, and both were limited to patients with acute hip fractures. The first included acute fracture surgery patients and omitted elective joint surgery patients, and it evaluated admission renal function but not postoperative AKI.²² The second study had a sample size of only 170 patients.²³ Thus, the literature leaves us with a crucial knowledge gap in renal outcomes and their postoperative impact in elective arthroplasties.

The present study filled this information gap by examining the incidence, risk factors, outcomes, and temporal trends of AKI after elective hip and knee arthroplasties. The increasing incidence of AKI in this surgical setting is similar to that of AKI in other surgical settings (cardiac and noncardiac).²¹ Although our analysis was limited by lack of perioperative management data, patients undergoing elective joint arthroplasty can experience kidney dysfunction for several reasons, including volume depletion, postoperative sepsis, and influence of medications, such as nonsteroidal anti-inflammatory drugs (NSAIDs), especially in older patients with more comorbidities and a higher burden of CKD. Each of these factors can cause renal dysfunction in patients having orthopedic procedures.²⁴ Moreover, NSAID use among elective joint arthroplasty patients is likely higher because of an emphasis on multimodal analgesia, as recent randomized controlled trials have demonstrated the efficacy of NSAID use in controlling pain without increasing bleeding.^25-27 Our results also demonstrated that the absolute incidence of AKI after orthopedic surgery is relatively low. One possible explanation for this phenomenon is that the definitions used were based on ICD-9-CM codes that underestimate the true incidence of AKI.

Consistent with other studies, we found that certain key preoperative comorbid conditions and postoperative events were associated with higher AKI risk. We stratified the rate of AKI associated with each postoperative event (sepsis, acute MI, cardiac catheterization, need for transfusion) by DM/CKD comorbidity. CKD was associated with significantly higher AKI risk across all postoperative complications. This information may provide clinicians with bedside information that can be used to determine which patients may be at higher or lower risk for AKI.

Our analysis of patient outcomes revealed that, though AKI was relatively uncommon, it increased the risk for death during hospitalization more than 10-fold between 2002 and 2012. Although the adjusted OR of in-hospital mortality decreased over the decade studied, the concurrent increase in AKI incidence caused the attributable risk of death associated with AKI to essentially remain the same. This observation is consistent with recent reports from cardiac surgery settings.²¹ These data together suggest that ameliorating occurrences of AKI would decrease mortality and increase quality of care for patients undergoing elective joint surgeries.

We also examined the effect of AKI on resource use by studying LOS, costs, and risk for adverse discharge. Much as in other surgical settings, AKI increased both LOS and overall hospitalization costs. More important, AKI was associated with increased adverse discharge (discharge to long-term care or nursing homes). Although exact reasons are unclear, we can speculate that postoperative renal dysfunction precludes early rehabilitation, impeding desired functional outcome and disposition.^28,29 Given the projected increases in primary and revision hip and knee arthroplasties,⁵ these data predict that the impact of AKI on health outcomes will increase alarmingly in coming years.

There are limitations to our study. First, it was based on administrative data and lacked patient-level and laboratory data. As reported, the sensitivity of AKI codes remains moderate,³⁰ so the true burden may be higher than indicated here. As the definition of AKI was based on administrative coding, we also could not estimate severity, though previous studies have found that administrative codes typically capture a more severe form of disease.³¹ Another limitation is that, because the data were deidentified, we could not delineate the risk for recurrent AKI in repeated surgical procedures, though this cohort unlikely was large enough to qualitatively affect our results. The third limitation is that, though we used CCI to adjust for the comorbidity burden, we were unable to account for other unmeasured confounders associated with increased AKI incidence, such as specific medication use. In addition, given the lack of patient-level data, we could not analyze the specific factors responsible for AKI in the perioperative period. Nevertheless, the strengths of a nationally representative sample, such as large sample size and generalizability, outweigh these limitations.

Conclusion

AKI is potentially an important quality indicator of elective joint surgery, and reducing its incidence is therefore essential for quality improvement. Given that hip and knee arthroplasties are projected to increase exponentially, as is the burden of comorbid conditions in this population, postoperative AKI will continue to have an incremental impact on health and health care resources. Thus, a carefully planned approach of interdisciplinary perioperative care is warranted to reduce both the risk and the consequences of this devastating condition.

Degenerative arthritis is a widespread chronic condition with an incidence of almost 43 million and annual health care costs of $60 billion in the United States alone.1 Although many cases can be managed symptomatically with medical therapy and intra-articular injections,2 many patients experience disease progression resulting in decreased ambulatory ability and work productivity. For these patients, elective hip and knee arthroplasties can drastically improve quality of life and functionality.3,4 Over the past decade, there has been a marked increase in the number of primary and revision total hip and knee arthroplasties performed in the United States. By 2030, the demand for primary total hip arthroplasties will grow an estimated 174%, to 572,000 procedures. Likewise, the demand for primary total knee arthroplasties is projected to grow by 673%, to 3.48 million procedures.5 However, though better surgical techniques and technology have led to improved functional outcomes, there is still substantial risk for complications in the perioperative period, especially in the geriatric population, in which substantial comorbidities are common.6-9

Acute kidney injury (AKI) is a common public health problem in hospitalized patients and in patients undergoing procedures. More than one-third of all AKI cases occur in surgical settings.^10,11 Over the past decade, both community-acquired and in-hospital AKIs rapidly increased in incidence in all major clinical settings.^12-14 Patients with AKI have high rates of adverse outcomes during hospitalization and discharge.^11,15 Sequelae of AKIs include worsening chronic kidney disease (CKD) and progression to end-stage renal disease, necessitating either long-term dialysis or transplantation.¹² This in turn leads to exacerbated disability, diminished quality of life, and disproportionate burden on health care resources.

Much of our knowledge about postoperative AKI has been derived from cardiovascular, thoracic, and abdominal surgery settings. However, there is a paucity of data on epidemiology and trends for either AKI or associated outcomes in patients undergoing major orthopedic surgery. The few studies to date either were single-center or had inadequate sample sizes for appropriately powered analysis of the risk factors and outcomes related to AKI.¹⁶

In the study reported here, we analyzed a large cohort of patients from a nationwide multicenter database to determine the incidence of and risk factors for AKI. We also examined the mortality and adverse discharges associated with AKI after major joint surgery. Lastly, we assessed temporal trends in both incidence and outcomes of AKI, including the death risk attributable to AKI.

Methods

Database

We extracted our study cohort from the Nationwide Inpatient Sample (NIS) and the National Inpatient Sample of Healthcare Cost and Utilization Project (HCUP) compiled by the Agency for Healthcare Research and Quality.¹⁷ NIS, the largest inpatient care database in the United States, stores data from almost 8 million stays in about 1000 hospitals across the country each year. Its participating hospital pool consists of about 20% of US community hospitals, resulting in a sampling frame comprising about 90% of all hospital discharges in the United States. This allows for calculation of precise, weighted nationwide estimates. Data elements within NIS are drawn from hospital discharge abstracts that indicate all procedures performed. NIS also stores information on patient characteristics, length of stay (LOS), discharge disposition, postoperative morbidity, and observed in-hospital mortality. However, it stores no information on long-term follow-up or complications after discharge.

Data Analysis

For the period 2002–2012, we queried the NIS database for hip and knee arthroplasties with primary diagnosis codes for osteoarthritis and secondary codes for AKI. We excluded patients under age 18 years and patients with diagnosis codes for hip and knee fracture/necrosis, inflammatory/infectious arthritis, or bone neoplasms (Table 1). We then extracted baseline characteristics of the study population. Patient-level characteristics included age, sex, race, quartile classification of median household income according to postal (ZIP) code, and primary payer (Medicare/Medicaid, private insurance, self-pay, no charge). Hospital-level characteristics included hospital location (urban, rural), hospital bed size (small, medium, large), region (Northeast, Midwest/North Central, South, West), and teaching status. We defined illness severity and likelihood of death using Deyo’s modification of the Charlson Comorbidity Index (CCI), which draws on principal and secondary ICD-9-CM (International Classification of Diseases, Ninth Revision-Clinical Modification) diagnosis codes, procedure codes, and patient demographics to estimate a patient’s mortality risk. This method reliably predicts mortality and readmission in the orthopedic population.^18,19 We assessed the effect of AKI on 4 outcomes, including in-hospital mortality, discharge disposition, LOS, and cost of stay. Discharge disposition was grouped by either (a) home or short-term facility or (b) adverse discharge. Home or short-term facility covered routine, short-term hospital, against medical advice, home intravenous provider, another rehabilitation facility, another institution for outpatient services, institution for outpatient services, discharged alive, and destination unknown; adverse discharge covered skilled nursing facility, intermediate care, hospice home, hospice medical facility, long-term care hospital, and certified nursing facility. This dichotomization of discharge disposition is often used in studies of NIS data.²⁰

Statistical Analyses

We compared the baseline characteristics of hospitalized patients with and without AKI. To test for significance, we used the χ² test for categorical variables, the Student t test for normally distributed continuous variables, the Wilcoxon rank sum test for non-normally distributed continuous variables, and the Cochran-Armitage test for trends in AKI incidence. We used survey logistic regression models to calculate adjusted odds ratios (ORs) with 95% confidence intervals (95% CIs) in order to estimate the predictors of AKI and the impact of AKI on hospital outcomes. We constructed final models after adjusting for confounders, testing for potential interactions, and ensuring no multicolinearity between covariates. Last, we computed the risk proportion of death attributable to AKI, indicating the proportion of deaths that could potentially be avoided if AKI and its complications were abrogated.²¹

We performed all statistical analyses with SAS Version 9.3 (SAS Institute) using designated weight values to produce weighted national estimates. The threshold for statistical significance was set at P < .01 (with ORs and 95% CIs that excluded 1).

Results

AKI Incidence, Risk Factors, and Trends

We identified 7,235,251 patients who underwent elective hip or knee arthroplasty for osteoarthritis between 2002 and 2012—an estimate consistent with data from the Centers for Disease Control and Prevention.²² Of that total, 94,367 (1.3%) had AKI. The proportion of discharges diagnosed with AKI increased rapidly over the decade, from 0.5% in 2002 to 1.8% to 1.9% in the period 2010–2012. This upward trend was highly significant (P_trend < .001) (Figure 1). Patients with AKI (vs patients without AKI) were more likely to be older (mean age, 70 vs 66 years; P < .001), male (50.8% vs 38.4%; P < .001), and black (10.07% vs 5.15%; P<. 001). They were also found to have a significantly higher comorbidity score (mean CCI, 2.8 vs 1.5; P < .001) and higher proportions of comorbidities, including hypertension, CKD, atrial fibrillation, diabetes mellitus (DM), congestive heart failure, chronic liver disease, and hepatitis C virus infection. In addition, AKI was associated with perioperative myocardial infarction (MI), sepsis, cardiac catheterization, and blood transfusion. Regarding socioeconomic characteristics, patients with AKI were more likely to have Medicare/Medicaid insurance (72.26% vs 58.06%; P < .001) and to belong to the extremes of income categories (Table 2).

Using multivariable logistic regression, we found that increased age (1.11 increase in adjusted OR for every year older; 95% CI, 1.09-1.14; P < .001), male sex (adjusted OR, 1.65; 95% CI, 1.60-1.71; P < .001), and black race (adjusted OR, 1.57; 95% CI, 1.45-1.69; P < .001) were significantly associated with postoperative AKI. Regarding comorbidities, baseline CKD (adjusted OR, 8.64; 95% CI, 8.14-9.18; P < .001) and congestive heart failure (adjusted OR, 2.74; 95% CI, 2.57-2.92; P< .0001) were most significantly associated with AKI. Perioperative events, including sepsis (adjusted OR, 35.64; 95% CI, 30.28-41.96; P < .0001), MI (adjusted OR, 6.14; 95% CI, 5.17-7.28; P < .0001), and blood transfusion (adjusted OR, 2.28; 95% CI, 2.15-2.42; P < .0001), were also strongly associated with postoperative AKI. Last, compared with urban hospitals and small hospital bed size, rural hospitals (adjusted OR, 0.70; 95% CI, 0.60-0.81; P< .001) and large bed size (adjusted OR, 0.82; 95% CI, 0.70-0.93; P = .003) were associated with lower probability of developing AKI (Table 3).

Figure 2 elucidates the frequency of AKI based on a combination of key preoperative comorbid conditions and postoperative complications—demonstrating that the proportion of AKI cases associated with other postoperative complications is significantly higher in the CKD and concomitant DM/CKD patient populations. Patients hospitalized with CKD exhibited higher rates of AKI in cases involving blood transfusion (20.9% vs 1.8%; P < .001), acute MI (48.9% vs 13.8%; P < .001), and sepsis (74.7% vs 36.3%;P< .001) relative to patients without CKD. Similarly, patients with concomitant DM/CKD exhibited higher rates of AKI in cases involving blood transfusion (23% vs 1.9%; P< .001), acute MI (51.1% vs 12.1%; P< .001), and sepsis (75% vs 38.2%; P < .001) relative to patients without either condition. However, patients hospitalized with DM alone exhibited only marginally higher rates of AKI in cases involving blood transfusion (4.7% vs 2%; P < .01) and acute MI (19.2% vs 16.7%; P< .01) and a lower rate in cases involving sepsis (38.2% vs 41.7%; P < .01) relative to patients without DM. These data suggest that CKD is the most significant clinically relevant risk factor for AKI and that CKD may synergize with DM to raise the risk for AKI.

Outcomes

We then analyzed the impact of AKI on hospital outcomes, including in-hospital mortality, discharge disposition, LOS, and cost of care. Mortality was significantly higher in patients with AKI than in patients without it (2.08% vs 0.06%; P < .001). Even after adjusting for confounders (eg, demographics, comorbidity burden, perioperative sepsis, hospital-level characteristics), AKI was still associated with strikingly higher odds of in-hospital death (adjusted OR, 11.32; 95% CI, 9.34-13.74; P < .001). However, analysis of temporal trends indicated that the odds for adjusted mortality associated with AKI decreased from 18.09 to 9.45 (P_trend = .01) over the period 2002–2012 (Figure 3). This decrease in odds of death was countered by an increase in incidence of AKI, resulting in a stable attributable risk proportion (97.9% in 2002 to 97.3% in 2012; P_trend = .90) (Table 4). Regarding discharge disposition, patients with AKI were much less likely to be discharged home (41.35% vs 62.59%; P < .001) and more likely to be discharged to long-term care (56.37% vs 37.03%; P< .001). After adjustment for confounders, AKI was associated with significantly increased odds of adverse discharge (adjusted OR, 2.24; 95% CI, 2.12-2.36; P< .001). Analysis of temporal trends revealed no appreciable decrease in the adjusted odds of adverse discharge between 2002 (adjusted OR, 1.87; 95% CI, 1.37-2.55; P < .001) and 2012 (adjusted OR, 1.93; 95% CI, 1.76-2.11; P < .001) (Figure 4, Table 5). Last, both mean LOS (5 days vs 3 days; P < .001) and mean cost of hospitalization (US $22,269 vs $15,757; P < .001) were significantly higher in patients with AKI.

Discussion

In this study, we found that the incidence of AKI among hospitalized patients increased 4-fold between 2002 and 2012. Moreover, we identified numerous patient-specific, hospital-specific, perioperative risk factors for AKI. Most important, we found that AKI was associated with a strikingly higher risk of in-hospital death, and surviving patients were more likely to experience adverse discharge. Although the adjusted mortality rate associated with AKI decreased over that decade, the attributable risk proportion remained stable.

Few studies have addressed this significant public health concern. In one recent study in Australia, Kimmel and colleagues¹⁶ identified risk factors for AKI but lacked data on AKI outcomes. In a study of complications and mortality occurring after orthopedic surgery, Belmont and colleagues²² categorized complications as either local or systemic but did not examine renal complications. Only 2 other major studies have been conducted on renal outcomes associated with major joint surgery, and both were limited to patients with acute hip fractures. The first included acute fracture surgery patients and omitted elective joint surgery patients, and it evaluated admission renal function but not postoperative AKI.²² The second study had a sample size of only 170 patients.²³ Thus, the literature leaves us with a crucial knowledge gap in renal outcomes and their postoperative impact in elective arthroplasties.

The present study filled this information gap by examining the incidence, risk factors, outcomes, and temporal trends of AKI after elective hip and knee arthroplasties. The increasing incidence of AKI in this surgical setting is similar to that of AKI in other surgical settings (cardiac and noncardiac).²¹ Although our analysis was limited by lack of perioperative management data, patients undergoing elective joint arthroplasty can experience kidney dysfunction for several reasons, including volume depletion, postoperative sepsis, and influence of medications, such as nonsteroidal anti-inflammatory drugs (NSAIDs), especially in older patients with more comorbidities and a higher burden of CKD. Each of these factors can cause renal dysfunction in patients having orthopedic procedures.²⁴ Moreover, NSAID use among elective joint arthroplasty patients is likely higher because of an emphasis on multimodal analgesia, as recent randomized controlled trials have demonstrated the efficacy of NSAID use in controlling pain without increasing bleeding.^25-27 Our results also demonstrated that the absolute incidence of AKI after orthopedic surgery is relatively low. One possible explanation for this phenomenon is that the definitions used were based on ICD-9-CM codes that underestimate the true incidence of AKI.

Consistent with other studies, we found that certain key preoperative comorbid conditions and postoperative events were associated with higher AKI risk. We stratified the rate of AKI associated with each postoperative event (sepsis, acute MI, cardiac catheterization, need for transfusion) by DM/CKD comorbidity. CKD was associated with significantly higher AKI risk across all postoperative complications. This information may provide clinicians with bedside information that can be used to determine which patients may be at higher or lower risk for AKI.

Our analysis of patient outcomes revealed that, though AKI was relatively uncommon, it increased the risk for death during hospitalization more than 10-fold between 2002 and 2012. Although the adjusted OR of in-hospital mortality decreased over the decade studied, the concurrent increase in AKI incidence caused the attributable risk of death associated with AKI to essentially remain the same. This observation is consistent with recent reports from cardiac surgery settings.²¹ These data together suggest that ameliorating occurrences of AKI would decrease mortality and increase quality of care for patients undergoing elective joint surgeries.

We also examined the effect of AKI on resource use by studying LOS, costs, and risk for adverse discharge. Much as in other surgical settings, AKI increased both LOS and overall hospitalization costs. More important, AKI was associated with increased adverse discharge (discharge to long-term care or nursing homes). Although exact reasons are unclear, we can speculate that postoperative renal dysfunction precludes early rehabilitation, impeding desired functional outcome and disposition.^28,29 Given the projected increases in primary and revision hip and knee arthroplasties,⁵ these data predict that the impact of AKI on health outcomes will increase alarmingly in coming years.

There are limitations to our study. First, it was based on administrative data and lacked patient-level and laboratory data. As reported, the sensitivity of AKI codes remains moderate,³⁰ so the true burden may be higher than indicated here. As the definition of AKI was based on administrative coding, we also could not estimate severity, though previous studies have found that administrative codes typically capture a more severe form of disease.³¹ Another limitation is that, because the data were deidentified, we could not delineate the risk for recurrent AKI in repeated surgical procedures, though this cohort unlikely was large enough to qualitatively affect our results. The third limitation is that, though we used CCI to adjust for the comorbidity burden, we were unable to account for other unmeasured confounders associated with increased AKI incidence, such as specific medication use. In addition, given the lack of patient-level data, we could not analyze the specific factors responsible for AKI in the perioperative period. Nevertheless, the strengths of a nationally representative sample, such as large sample size and generalizability, outweigh these limitations.

Conclusion

AKI is potentially an important quality indicator of elective joint surgery, and reducing its incidence is therefore essential for quality improvement. Given that hip and knee arthroplasties are projected to increase exponentially, as is the burden of comorbid conditions in this population, postoperative AKI will continue to have an incremental impact on health and health care resources. Thus, a carefully planned approach of interdisciplinary perioperative care is warranted to reduce both the risk and the consequences of this devastating condition.

Musculoskeletal disorders, the leading cause of disability in the United States,¹ account for more than half of all persons reporting missing a workday because of a medical condition.² Shoulder disorders in particular play a significant role in the burden of musculoskeletal disorders and cost of care. In 2008, 18.9 million adults (8.2% of the US adult population) reported chronic shoulder pain.¹ Among shoulder disorders, rotator cuff pathology is the most common cause of shoulder-related disability found by orthopedic surgeons.³ Rotator cuff surgery (RCS) is one of the most commonly performed orthopedic surgical procedures, and surgery volume is on the rise. One study found a 141% increase in rotator cuff repairs between the years 1996 (~41 per 100,000 population) and 2006 (~98 per 100,000 population).⁴

US health care costs are also increasing. In 2011, $2.7 trillion was spent on health care, representing 17.9% of the national gross domestic product (GDP). According to projections, costs will rise to $4.6 trillion by 2020.5 In particular, as patients continue to live longer and remain more active into their later years, the costs of treating and managing musculoskeletal disorders become more important from a public policy standpoint. In 2006, the cost of treating musculoskeletal disorders alone was $576 billion, representing 4.5% of that year’s GDP.²

Paramount in this era of rising costs is the idea of maximizing the value of health care dollars. Health care economists Porter and Teisberg⁶ defined value as patient health outcomes achieved per dollar of cost expended in a care cycle (diagnosis, treatment, ongoing management) for a particular disease or disorder. For proper management of value, outcomes and costs for an entire cycle of care must be determined. From a practical standpoint, this first requires determining the true cost of a care cycle—dollars spent on personnel, equipment, materials, and other resources required to deliver a particular service—rather than the amount charged or reimbursed for providing the service in question.⁷

Kaplan and Anderson^8,9 described the TDABC (time-driven activity-based costing) algorithm for calculating the cost of delivering a service based on 2 parameters: unit cost of a particular resource, and time required to supply it. These parameters apply to material costs and labor costs. In the medical setting, the TDABC algorithm can be applied by defining a care delivery value chain for each aspect of patient care and then multiplying incremental cost per unit time by time required to deliver that resource (Figure 1). Tabulating the overall unit cost for each resource then yields the overall cost of the care cycle. Clinical outcomes data can then be determined and used to calculate overall value for the patient care cycle.

In the study reported here, we used the TDABC algorithm to calculate the direct financial costs of surgical treatment of rotator cuff tears confirmed by magnetic resonance imaging (MRI) in an academic medical center.

Methods

Per our institution’s Office for the Protection of Research Subjects, institutional review board (IRB) approval is required only for projects using “human subjects” as defined by federal policy. In the present study, no private information could be identified, and all data were obtained from hospital billing records without intervention or interaction with individual patients. Accordingly, IRB approval was deemed unnecessary for our economic cost analysis.

Billing records of a single academic fellowship-trained sports surgeon were reviewed to identify patients who underwent primary repair of an MRI-confirmed rotator cuff tear between April 1, 2009, and July 31, 2012. Patients who had undergone prior shoulder surgery of any type were excluded from the study. Operative reports were reviewed, and exact surgical procedures performed were noted. The operating surgeon selected the specific repair techniques, including single- or double-row repair, with emphasis on restoring footprint coverage and avoiding overtensioning.

All surgeries were performed in an outpatient surgical center owned and operated by the surgeon’s home university. Surgeries were performed by the attending physician assisted by a senior orthopedic resident. The RCS care cycle was divided into 3 phases (Figure 2):

1. Preoperative. Patient’s interaction with receptionist in surgery center, time with preoperative nurse and circulating nurse in preoperative area, resident check-in time, and time placing preoperative nerve block and consumable materials used during block placement.

2. Operative. Time in operating room with surgical team for RCS, consumable materials used during surgery (eg, anchors, shavers, drapes), anesthetic medications, shoulder abduction pillow placed on completion of surgery, and cost of instrument processing.

3. Postoperative. Time in postoperative recovery area with recovery room nursing staff.

Time in each portion of the care cycle was directly observed and tabulated by hospital volunteers in the surgery center. Institutional billing data were used to identify material resources consumed, and the actual cost paid by the hospital for these resources was obtained from internal records. Mean hourly salary data and standard benefit rates were obtained for surgery center staff. Attending physician salary was extrapolated from published mean market salary data for academic physicians and mean hours worked,^10,11 and resident physician costs were tabulated from publically available institutional payroll data and average resident work hours at our institution. These cost data and times were then used to tabulate total cost for the RCS care cycle using the TDABC algorithm.

Results

We identified 28 shoulders in 26 patients (mean age, 54.5 years) who met the inclusion criteria. Of these 28 shoulders, 18 (64.3%) had an isolated supraspinatus tear, 8 (28.6%) had combined supraspinatus and infraspinatus tears, 1 (3.6%) had combined supraspinatus and subscapularis tears, and 1 (3.6%) had an isolated infraspinatus tear. Demographic data are listed in Table 1.

All patients received an interscalene nerve block in the preoperative area before being brought into the operating room. In our analysis, we included nerve block supply costs and the anesthesiologist’s mean time placing the nerve block.

In all cases, primary rotator cuff repair was performed with suture anchors (Parcus Medical) with the patient in the lateral decubitus position. In 13 (46%) of the 28 shoulders, this repair was described as “complex,” requiring double-row technique. Subacromial decompression and bursectomy were performed in addition to the rotator cuff repair. Labral débridement was performed in 23 patients, synovectomy in 10, biceps tenodesis with anchor (Smith & Nephew) in 1, and biceps tenotomy in 1. Mean time in operating room was 148 minutes; mean time in postoperative recovery unit was 105 minutes.

Directly observing the care cycle, hospital volunteers found that patients spent a mean of 15 minutes with the receptionist when they arrived in the outpatient surgical center, 25 minutes with nurses for check-in in the preoperative holding area, and 10 minutes with the anesthesiology resident and 15 minutes with the orthopedic surgery resident for preoperative evaluation and paperwork. Mean nerve block time was 20 minutes. Mean electrocardiogram (ECG) time (12 patients) was 15 minutes. The surgical technician spent a mean time of 20 minutes setting up the operating room before the patient was brought in and 15 minutes cleaning up after the patient was transferred to the recovery room. Costs of postoperative care in the recovery room were based on a 2:1 patient-to-nurse ratio, as is the standard practice in our outpatient surgery center.

Using the times mentioned and our hospital’s salary data—including standard hospital benefits rates of 33.5% for nonphysicians and 17.65% for physicians—we determined, using the TDABC algorithm, a direct cost of $5904.21 for this process cycle, excluding hospital overhead and indirect costs. Table 2 provides the overall cost breakdown. Compared with the direct economic cost, the mean hospital charge to insurers for the procedure was $31,459.35. Mean reimbursement from insurers was $9679.08.

Overall attending and resident physician costs were $1077.75, which consisted of $623.66 for the surgeon and $454.09 for the anesthesiologist (included placement of nerve block and administration of anesthesia during surgery). Preoperative bloodwork was obtained in 23 cases, adding a mean cost of $111.04 after adjusting for standard hospital markup. Preoperative ECG was performed in 12 cases, for an added mean cost of $7.30 based on the TDABC algorithm.

We also broke down costs by care cycle phase. The preoperative phase, excluding the preoperative laboratory studies and ECGs (not performed in all cases), cost $134.34 (2.3% of total costs); the operative phase cost $5718.01 (96.8% of total costs); and the postoperative phase cost $51.86 (0.9% of total costs). Within the operative phase, the cost of consumables (specifically, suture anchors) was the main cost driver. Mean anchor cost per case was $3432.67. “Complex” tears involving a double-row repair averaged $4570.25 in anchor cost per patient, as compared with $2522.60 in anchor costs for simple repairs.

Discussion

US health care costs continue to increase unsustainably, with rising pressure on hospitals and providers to deliver the highest value for each health care dollar. The present study is the first to calculate (using the TDABC algorithm) the direct economic cost ($5904.21) of the entire RCS care cycle at a university-based outpatient surgery center. Rent, utility costs, administrative costs, overhead, and other indirect costs at the surgery center were not included in this cost analysis, as they would be incurred irrespective of type of surgery performed. As such, our data isolate the procedure-specific costs of rotator cuff repair in order to provide a more meaningful comparison for other institutions, where indirect costs may be different.

In the literature, rigorous economic analysis of shoulder pathology is sparse. Kuye and colleagues¹² systematically reviewed economic evaluations in shoulder surgery for the period 1980–2010 and noted more than 50% of the papers were published between 2005 and 2010.¹² They also noted the poor quality of these studies and concluded more rigorous economic evaluations are needed to help justify the rising costs of shoulder-related treatments.

Several studies have directly evaluated costs associated with RCS. Cordasco and colleagues¹³ detailed the success of open rotator cuff repair as an outpatient procedure—noting its 43% cost savings ($4300 for outpatient vs $7500 for inpatient) and high patient satisfaction—using hospital charge data for operating room time, supplies, instruments, and postoperative slings. Churchill and Ghorai¹⁴ evaluated costs of mini-open and arthroscopic rotator cuff repairs in a statewide database and estimated the arthroscopic repair cost at $8985, compared with $7841 for the mini-open repair. They used reported hospital charge data, which were not itemized and did not include physician professional fees. Adla and colleagues,¹⁵ in a similar analysis of open versus arthroscopic cuff repair, estimated direct material costs of $1609.50 (arthroscopic) and $360.75 (open); these figures were converted from 2005 UK currency using the exchange rate cited in their paper. Salaries of surgeon, anesthesiologist, and other operating room personnel were said to be included in the operating room cost, but the authors’ paper did not include these data.

Two studies directly estimated the costs of arthroscopic rotator cuff repair. Hearnden and Tennent¹⁶ calculated the cost of RCS at their UK institution to be £2672, which included cost of operating room consumable materials, medication, and salaries of operating room personnel, including surgeon and anesthesiologist. Using online currency conversion from 2008 exchange rates and adjusting for inflation gave a corresponding US cost of $5449.63.¹⁷ Vitale and colleagues¹⁸ prospectively calculated costs of arthroscopic rotator cuff repair over a 1-year period using a cost-to-charge ratio from tabulated inpatient charges, procedure charges, and physician fees and payments abstracted from medical records, hospital billing, and administrative databases. Mean total cost for this cycle was $10,605.20, which included several costs (physical therapy, radiologist fees) not included in the present study. These studies, though more comprehensive than prior work, did not capture the entire cycle of surgical care.

Our study was designed to provide initial data on the direct costs of arthroscopic repair of the rotator cuff for the entire process cycle. Our overall cost estimate of $5904.21 differs significantly from prior work—not unexpected given the completely different cost methodology used.

Our study had several limitations. First, it was a single-surgeon evaluation, and a number of operating room variables (eg, use of adjunct instrumentation such as radiofrequency probes, differences in draping preferences) as well as surgeon volume in performing rotator cuff repairs might have substantially affected the reproducibility and generalizability of our data. Similarly, the large number of adjunctive procedures (eg, subacromial decompression, labral débridement) performed in conjunction with the rotator cuff repairs added operative time and therefore increased overall cost. Double-row repairs added operative time and increased the cost of consumable materials as well. Differences in surgeon preference for suture anchors may also be important, as anchors are a major cost driver and can vary significantly between vendors and institutions. Tear-related variables (eg, tear size, tear chronicity, degree of fatty cuff degeneration) were not controlled for and might have significantly affected operative time and associated cost. Resident involvement in the surgical procedure and anesthesia process in an academic setting prolongs surgical time and thus directly impacts costs.

In addition, we used the patient’s time in the operating room as a proxy for actual surgical time, as this was the only reliable and reproducible data point available in our electronic medical record. As such, an unquantifiable amount of surgeon time may have been overallocated to our cost estimate for time spent inducing anesthesia, positioning, helping take the patient off the operating table, and so on. However, as typical surgeon practice is to be involved in these tasks in the operating room, the possible overestimate of surgeon cost is likely minimal.

Our salary data for the TDABC algorithm were based on national averages for work hours and gross income for physicians and on hospital-based wage structure and may not be generalizable to other institutions. There may also be regional differences in work hours and salaries, which in turn would factor into a different per-minute cost for surgeon and anesthesiologist, depending on the exact geographic area where the surgery is performed. Costs may be higher at institutions that use certified nurse anesthetists rather than resident physicians because of the salary differences between these practitioners.

Moreover, the time that patients spend in the holding area—waiting to go into surgery and, after surgery, waiting for their ride home, for their prescriptions to be ready, and so forth—is an important variable to consider from a cost standpoint. However, as this time varied significantly and involved minimal contact with hospital personnel, we excluded its associated costs from our analysis. Similarly, and as already noted, hospital overhead and other indirect costs were excluded from analysis as well.

Conclusion

Using the TDABC algorithm, we found a direct economic cost of $5904.21 for RCS at our academic outpatient surgical center, with anchor cost the main cost driver. Judicious use of consumable resources is a key focus for cost containment in arthroscopic shoulder surgery, particularly with respect to implantable suture anchors. However, in the setting of more complex tears that require multiple anchors in a double-row repair construct, our pilot data may be useful to hospitals and surgery centers negotiating procedural reimbursement for the increased cost of complex repairs. Use of the TDABC algorithm for RCS and other procedures may also help in identifying opportunities to deliver more cost-effective health care.

Musculoskeletal disorders, the leading cause of disability in the United States,¹ account for more than half of all persons reporting missing a workday because of a medical condition.² Shoulder disorders in particular play a significant role in the burden of musculoskeletal disorders and cost of care. In 2008, 18.9 million adults (8.2% of the US adult population) reported chronic shoulder pain.¹ Among shoulder disorders, rotator cuff pathology is the most common cause of shoulder-related disability found by orthopedic surgeons.³ Rotator cuff surgery (RCS) is one of the most commonly performed orthopedic surgical procedures, and surgery volume is on the rise. One study found a 141% increase in rotator cuff repairs between the years 1996 (~41 per 100,000 population) and 2006 (~98 per 100,000 population).⁴

US health care costs are also increasing. In 2011, $2.7 trillion was spent on health care, representing 17.9% of the national gross domestic product (GDP). According to projections, costs will rise to $4.6 trillion by 2020.5 In particular, as patients continue to live longer and remain more active into their later years, the costs of treating and managing musculoskeletal disorders become more important from a public policy standpoint. In 2006, the cost of treating musculoskeletal disorders alone was $576 billion, representing 4.5% of that year’s GDP.²

Paramount in this era of rising costs is the idea of maximizing the value of health care dollars. Health care economists Porter and Teisberg⁶ defined value as patient health outcomes achieved per dollar of cost expended in a care cycle (diagnosis, treatment, ongoing management) for a particular disease or disorder. For proper management of value, outcomes and costs for an entire cycle of care must be determined. From a practical standpoint, this first requires determining the true cost of a care cycle—dollars spent on personnel, equipment, materials, and other resources required to deliver a particular service—rather than the amount charged or reimbursed for providing the service in question.⁷

Kaplan and Anderson^8,9 described the TDABC (time-driven activity-based costing) algorithm for calculating the cost of delivering a service based on 2 parameters: unit cost of a particular resource, and time required to supply it. These parameters apply to material costs and labor costs. In the medical setting, the TDABC algorithm can be applied by defining a care delivery value chain for each aspect of patient care and then multiplying incremental cost per unit time by time required to deliver that resource (Figure 1). Tabulating the overall unit cost for each resource then yields the overall cost of the care cycle. Clinical outcomes data can then be determined and used to calculate overall value for the patient care cycle.

In the study reported here, we used the TDABC algorithm to calculate the direct financial costs of surgical treatment of rotator cuff tears confirmed by magnetic resonance imaging (MRI) in an academic medical center.

Methods

Per our institution’s Office for the Protection of Research Subjects, institutional review board (IRB) approval is required only for projects using “human subjects” as defined by federal policy. In the present study, no private information could be identified, and all data were obtained from hospital billing records without intervention or interaction with individual patients. Accordingly, IRB approval was deemed unnecessary for our economic cost analysis.

Billing records of a single academic fellowship-trained sports surgeon were reviewed to identify patients who underwent primary repair of an MRI-confirmed rotator cuff tear between April 1, 2009, and July 31, 2012. Patients who had undergone prior shoulder surgery of any type were excluded from the study. Operative reports were reviewed, and exact surgical procedures performed were noted. The operating surgeon selected the specific repair techniques, including single- or double-row repair, with emphasis on restoring footprint coverage and avoiding overtensioning.

All surgeries were performed in an outpatient surgical center owned and operated by the surgeon’s home university. Surgeries were performed by the attending physician assisted by a senior orthopedic resident. The RCS care cycle was divided into 3 phases (Figure 2):

1. Preoperative. Patient’s interaction with receptionist in surgery center, time with preoperative nurse and circulating nurse in preoperative area, resident check-in time, and time placing preoperative nerve block and consumable materials used during block placement.

2. Operative. Time in operating room with surgical team for RCS, consumable materials used during surgery (eg, anchors, shavers, drapes), anesthetic medications, shoulder abduction pillow placed on completion of surgery, and cost of instrument processing.

3. Postoperative. Time in postoperative recovery area with recovery room nursing staff.

Time in each portion of the care cycle was directly observed and tabulated by hospital volunteers in the surgery center. Institutional billing data were used to identify material resources consumed, and the actual cost paid by the hospital for these resources was obtained from internal records. Mean hourly salary data and standard benefit rates were obtained for surgery center staff. Attending physician salary was extrapolated from published mean market salary data for academic physicians and mean hours worked,^10,11 and resident physician costs were tabulated from publically available institutional payroll data and average resident work hours at our institution. These cost data and times were then used to tabulate total cost for the RCS care cycle using the TDABC algorithm.

Results

We identified 28 shoulders in 26 patients (mean age, 54.5 years) who met the inclusion criteria. Of these 28 shoulders, 18 (64.3%) had an isolated supraspinatus tear, 8 (28.6%) had combined supraspinatus and infraspinatus tears, 1 (3.6%) had combined supraspinatus and subscapularis tears, and 1 (3.6%) had an isolated infraspinatus tear. Demographic data are listed in Table 1.

All patients received an interscalene nerve block in the preoperative area before being brought into the operating room. In our analysis, we included nerve block supply costs and the anesthesiologist’s mean time placing the nerve block.

In all cases, primary rotator cuff repair was performed with suture anchors (Parcus Medical) with the patient in the lateral decubitus position. In 13 (46%) of the 28 shoulders, this repair was described as “complex,” requiring double-row technique. Subacromial decompression and bursectomy were performed in addition to the rotator cuff repair. Labral débridement was performed in 23 patients, synovectomy in 10, biceps tenodesis with anchor (Smith & Nephew) in 1, and biceps tenotomy in 1. Mean time in operating room was 148 minutes; mean time in postoperative recovery unit was 105 minutes.

Directly observing the care cycle, hospital volunteers found that patients spent a mean of 15 minutes with the receptionist when they arrived in the outpatient surgical center, 25 minutes with nurses for check-in in the preoperative holding area, and 10 minutes with the anesthesiology resident and 15 minutes with the orthopedic surgery resident for preoperative evaluation and paperwork. Mean nerve block time was 20 minutes. Mean electrocardiogram (ECG) time (12 patients) was 15 minutes. The surgical technician spent a mean time of 20 minutes setting up the operating room before the patient was brought in and 15 minutes cleaning up after the patient was transferred to the recovery room. Costs of postoperative care in the recovery room were based on a 2:1 patient-to-nurse ratio, as is the standard practice in our outpatient surgery center.

Using the times mentioned and our hospital’s salary data—including standard hospital benefits rates of 33.5% for nonphysicians and 17.65% for physicians—we determined, using the TDABC algorithm, a direct cost of $5904.21 for this process cycle, excluding hospital overhead and indirect costs. Table 2 provides the overall cost breakdown. Compared with the direct economic cost, the mean hospital charge to insurers for the procedure was $31,459.35. Mean reimbursement from insurers was $9679.08.

Overall attending and resident physician costs were $1077.75, which consisted of $623.66 for the surgeon and $454.09 for the anesthesiologist (included placement of nerve block and administration of anesthesia during surgery). Preoperative bloodwork was obtained in 23 cases, adding a mean cost of $111.04 after adjusting for standard hospital markup. Preoperative ECG was performed in 12 cases, for an added mean cost of $7.30 based on the TDABC algorithm.

We also broke down costs by care cycle phase. The preoperative phase, excluding the preoperative laboratory studies and ECGs (not performed in all cases), cost $134.34 (2.3% of total costs); the operative phase cost $5718.01 (96.8% of total costs); and the postoperative phase cost $51.86 (0.9% of total costs). Within the operative phase, the cost of consumables (specifically, suture anchors) was the main cost driver. Mean anchor cost per case was $3432.67. “Complex” tears involving a double-row repair averaged $4570.25 in anchor cost per patient, as compared with $2522.60 in anchor costs for simple repairs.

Discussion

US health care costs continue to increase unsustainably, with rising pressure on hospitals and providers to deliver the highest value for each health care dollar. The present study is the first to calculate (using the TDABC algorithm) the direct economic cost ($5904.21) of the entire RCS care cycle at a university-based outpatient surgery center. Rent, utility costs, administrative costs, overhead, and other indirect costs at the surgery center were not included in this cost analysis, as they would be incurred irrespective of type of surgery performed. As such, our data isolate the procedure-specific costs of rotator cuff repair in order to provide a more meaningful comparison for other institutions, where indirect costs may be different.

In the literature, rigorous economic analysis of shoulder pathology is sparse. Kuye and colleagues¹² systematically reviewed economic evaluations in shoulder surgery for the period 1980–2010 and noted more than 50% of the papers were published between 2005 and 2010.¹² They also noted the poor quality of these studies and concluded more rigorous economic evaluations are needed to help justify the rising costs of shoulder-related treatments.

Several studies have directly evaluated costs associated with RCS. Cordasco and colleagues¹³ detailed the success of open rotator cuff repair as an outpatient procedure—noting its 43% cost savings ($4300 for outpatient vs $7500 for inpatient) and high patient satisfaction—using hospital charge data for operating room time, supplies, instruments, and postoperative slings. Churchill and Ghorai¹⁴ evaluated costs of mini-open and arthroscopic rotator cuff repairs in a statewide database and estimated the arthroscopic repair cost at $8985, compared with $7841 for the mini-open repair. They used reported hospital charge data, which were not itemized and did not include physician professional fees. Adla and colleagues,¹⁵ in a similar analysis of open versus arthroscopic cuff repair, estimated direct material costs of $1609.50 (arthroscopic) and $360.75 (open); these figures were converted from 2005 UK currency using the exchange rate cited in their paper. Salaries of surgeon, anesthesiologist, and other operating room personnel were said to be included in the operating room cost, but the authors’ paper did not include these data.

Two studies directly estimated the costs of arthroscopic rotator cuff repair. Hearnden and Tennent¹⁶ calculated the cost of RCS at their UK institution to be £2672, which included cost of operating room consumable materials, medication, and salaries of operating room personnel, including surgeon and anesthesiologist. Using online currency conversion from 2008 exchange rates and adjusting for inflation gave a corresponding US cost of $5449.63.¹⁷ Vitale and colleagues¹⁸ prospectively calculated costs of arthroscopic rotator cuff repair over a 1-year period using a cost-to-charge ratio from tabulated inpatient charges, procedure charges, and physician fees and payments abstracted from medical records, hospital billing, and administrative databases. Mean total cost for this cycle was $10,605.20, which included several costs (physical therapy, radiologist fees) not included in the present study. These studies, though more comprehensive than prior work, did not capture the entire cycle of surgical care.

Our study was designed to provide initial data on the direct costs of arthroscopic repair of the rotator cuff for the entire process cycle. Our overall cost estimate of $5904.21 differs significantly from prior work—not unexpected given the completely different cost methodology used.

Our study had several limitations. First, it was a single-surgeon evaluation, and a number of operating room variables (eg, use of adjunct instrumentation such as radiofrequency probes, differences in draping preferences) as well as surgeon volume in performing rotator cuff repairs might have substantially affected the reproducibility and generalizability of our data. Similarly, the large number of adjunctive procedures (eg, subacromial decompression, labral débridement) performed in conjunction with the rotator cuff repairs added operative time and therefore increased overall cost. Double-row repairs added operative time and increased the cost of consumable materials as well. Differences in surgeon preference for suture anchors may also be important, as anchors are a major cost driver and can vary significantly between vendors and institutions. Tear-related variables (eg, tear size, tear chronicity, degree of fatty cuff degeneration) were not controlled for and might have significantly affected operative time and associated cost. Resident involvement in the surgical procedure and anesthesia process in an academic setting prolongs surgical time and thus directly impacts costs.

In addition, we used the patient’s time in the operating room as a proxy for actual surgical time, as this was the only reliable and reproducible data point available in our electronic medical record. As such, an unquantifiable amount of surgeon time may have been overallocated to our cost estimate for time spent inducing anesthesia, positioning, helping take the patient off the operating table, and so on. However, as typical surgeon practice is to be involved in these tasks in the operating room, the possible overestimate of surgeon cost is likely minimal.

Our salary data for the TDABC algorithm were based on national averages for work hours and gross income for physicians and on hospital-based wage structure and may not be generalizable to other institutions. There may also be regional differences in work hours and salaries, which in turn would factor into a different per-minute cost for surgeon and anesthesiologist, depending on the exact geographic area where the surgery is performed. Costs may be higher at institutions that use certified nurse anesthetists rather than resident physicians because of the salary differences between these practitioners.

Moreover, the time that patients spend in the holding area—waiting to go into surgery and, after surgery, waiting for their ride home, for their prescriptions to be ready, and so forth—is an important variable to consider from a cost standpoint. However, as this time varied significantly and involved minimal contact with hospital personnel, we excluded its associated costs from our analysis. Similarly, and as already noted, hospital overhead and other indirect costs were excluded from analysis as well.

Conclusion

Using the TDABC algorithm, we found a direct economic cost of $5904.21 for RCS at our academic outpatient surgical center, with anchor cost the main cost driver. Judicious use of consumable resources is a key focus for cost containment in arthroscopic shoulder surgery, particularly with respect to implantable suture anchors. However, in the setting of more complex tears that require multiple anchors in a double-row repair construct, our pilot data may be useful to hospitals and surgery centers negotiating procedural reimbursement for the increased cost of complex repairs. Use of the TDABC algorithm for RCS and other procedures may also help in identifying opportunities to deliver more cost-effective health care.

Musculoskeletal disorders, the leading cause of disability in the United States,¹ account for more than half of all persons reporting missing a workday because of a medical condition.² Shoulder disorders in particular play a significant role in the burden of musculoskeletal disorders and cost of care. In 2008, 18.9 million adults (8.2% of the US adult population) reported chronic shoulder pain.¹ Among shoulder disorders, rotator cuff pathology is the most common cause of shoulder-related disability found by orthopedic surgeons.³ Rotator cuff surgery (RCS) is one of the most commonly performed orthopedic surgical procedures, and surgery volume is on the rise. One study found a 141% increase in rotator cuff repairs between the years 1996 (~41 per 100,000 population) and 2006 (~98 per 100,000 population).⁴

US health care costs are also increasing. In 2011, $2.7 trillion was spent on health care, representing 17.9% of the national gross domestic product (GDP). According to projections, costs will rise to $4.6 trillion by 2020.5 In particular, as patients continue to live longer and remain more active into their later years, the costs of treating and managing musculoskeletal disorders become more important from a public policy standpoint. In 2006, the cost of treating musculoskeletal disorders alone was $576 billion, representing 4.5% of that year’s GDP.²

Paramount in this era of rising costs is the idea of maximizing the value of health care dollars. Health care economists Porter and Teisberg⁶ defined value as patient health outcomes achieved per dollar of cost expended in a care cycle (diagnosis, treatment, ongoing management) for a particular disease or disorder. For proper management of value, outcomes and costs for an entire cycle of care must be determined. From a practical standpoint, this first requires determining the true cost of a care cycle—dollars spent on personnel, equipment, materials, and other resources required to deliver a particular service—rather than the amount charged or reimbursed for providing the service in question.⁷

Kaplan and Anderson^8,9 described the TDABC (time-driven activity-based costing) algorithm for calculating the cost of delivering a service based on 2 parameters: unit cost of a particular resource, and time required to supply it. These parameters apply to material costs and labor costs. In the medical setting, the TDABC algorithm can be applied by defining a care delivery value chain for each aspect of patient care and then multiplying incremental cost per unit time by time required to deliver that resource (Figure 1). Tabulating the overall unit cost for each resource then yields the overall cost of the care cycle. Clinical outcomes data can then be determined and used to calculate overall value for the patient care cycle.

In the study reported here, we used the TDABC algorithm to calculate the direct financial costs of surgical treatment of rotator cuff tears confirmed by magnetic resonance imaging (MRI) in an academic medical center.

Methods

Per our institution’s Office for the Protection of Research Subjects, institutional review board (IRB) approval is required only for projects using “human subjects” as defined by federal policy. In the present study, no private information could be identified, and all data were obtained from hospital billing records without intervention or interaction with individual patients. Accordingly, IRB approval was deemed unnecessary for our economic cost analysis.

Billing records of a single academic fellowship-trained sports surgeon were reviewed to identify patients who underwent primary repair of an MRI-confirmed rotator cuff tear between April 1, 2009, and July 31, 2012. Patients who had undergone prior shoulder surgery of any type were excluded from the study. Operative reports were reviewed, and exact surgical procedures performed were noted. The operating surgeon selected the specific repair techniques, including single- or double-row repair, with emphasis on restoring footprint coverage and avoiding overtensioning.

All surgeries were performed in an outpatient surgical center owned and operated by the surgeon’s home university. Surgeries were performed by the attending physician assisted by a senior orthopedic resident. The RCS care cycle was divided into 3 phases (Figure 2):

1. Preoperative. Patient’s interaction with receptionist in surgery center, time with preoperative nurse and circulating nurse in preoperative area, resident check-in time, and time placing preoperative nerve block and consumable materials used during block placement.

2. Operative. Time in operating room with surgical team for RCS, consumable materials used during surgery (eg, anchors, shavers, drapes), anesthetic medications, shoulder abduction pillow placed on completion of surgery, and cost of instrument processing.

3. Postoperative. Time in postoperative recovery area with recovery room nursing staff.

Time in each portion of the care cycle was directly observed and tabulated by hospital volunteers in the surgery center. Institutional billing data were used to identify material resources consumed, and the actual cost paid by the hospital for these resources was obtained from internal records. Mean hourly salary data and standard benefit rates were obtained for surgery center staff. Attending physician salary was extrapolated from published mean market salary data for academic physicians and mean hours worked,^10,11 and resident physician costs were tabulated from publically available institutional payroll data and average resident work hours at our institution. These cost data and times were then used to tabulate total cost for the RCS care cycle using the TDABC algorithm.

Results

We identified 28 shoulders in 26 patients (mean age, 54.5 years) who met the inclusion criteria. Of these 28 shoulders, 18 (64.3%) had an isolated supraspinatus tear, 8 (28.6%) had combined supraspinatus and infraspinatus tears, 1 (3.6%) had combined supraspinatus and subscapularis tears, and 1 (3.6%) had an isolated infraspinatus tear. Demographic data are listed in Table 1.

All patients received an interscalene nerve block in the preoperative area before being brought into the operating room. In our analysis, we included nerve block supply costs and the anesthesiologist’s mean time placing the nerve block.

In all cases, primary rotator cuff repair was performed with suture anchors (Parcus Medical) with the patient in the lateral decubitus position. In 13 (46%) of the 28 shoulders, this repair was described as “complex,” requiring double-row technique. Subacromial decompression and bursectomy were performed in addition to the rotator cuff repair. Labral débridement was performed in 23 patients, synovectomy in 10, biceps tenodesis with anchor (Smith & Nephew) in 1, and biceps tenotomy in 1. Mean time in operating room was 148 minutes; mean time in postoperative recovery unit was 105 minutes.

Directly observing the care cycle, hospital volunteers found that patients spent a mean of 15 minutes with the receptionist when they arrived in the outpatient surgical center, 25 minutes with nurses for check-in in the preoperative holding area, and 10 minutes with the anesthesiology resident and 15 minutes with the orthopedic surgery resident for preoperative evaluation and paperwork. Mean nerve block time was 20 minutes. Mean electrocardiogram (ECG) time (12 patients) was 15 minutes. The surgical technician spent a mean time of 20 minutes setting up the operating room before the patient was brought in and 15 minutes cleaning up after the patient was transferred to the recovery room. Costs of postoperative care in the recovery room were based on a 2:1 patient-to-nurse ratio, as is the standard practice in our outpatient surgery center.

Using the times mentioned and our hospital’s salary data—including standard hospital benefits rates of 33.5% for nonphysicians and 17.65% for physicians—we determined, using the TDABC algorithm, a direct cost of $5904.21 for this process cycle, excluding hospital overhead and indirect costs. Table 2 provides the overall cost breakdown. Compared with the direct economic cost, the mean hospital charge to insurers for the procedure was $31,459.35. Mean reimbursement from insurers was $9679.08.

Overall attending and resident physician costs were $1077.75, which consisted of $623.66 for the surgeon and $454.09 for the anesthesiologist (included placement of nerve block and administration of anesthesia during surgery). Preoperative bloodwork was obtained in 23 cases, adding a mean cost of $111.04 after adjusting for standard hospital markup. Preoperative ECG was performed in 12 cases, for an added mean cost of $7.30 based on the TDABC algorithm.

We also broke down costs by care cycle phase. The preoperative phase, excluding the preoperative laboratory studies and ECGs (not performed in all cases), cost $134.34 (2.3% of total costs); the operative phase cost $5718.01 (96.8% of total costs); and the postoperative phase cost $51.86 (0.9% of total costs). Within the operative phase, the cost of consumables (specifically, suture anchors) was the main cost driver. Mean anchor cost per case was $3432.67. “Complex” tears involving a double-row repair averaged $4570.25 in anchor cost per patient, as compared with $2522.60 in anchor costs for simple repairs.

Discussion

US health care costs continue to increase unsustainably, with rising pressure on hospitals and providers to deliver the highest value for each health care dollar. The present study is the first to calculate (using the TDABC algorithm) the direct economic cost ($5904.21) of the entire RCS care cycle at a university-based outpatient surgery center. Rent, utility costs, administrative costs, overhead, and other indirect costs at the surgery center were not included in this cost analysis, as they would be incurred irrespective of type of surgery performed. As such, our data isolate the procedure-specific costs of rotator cuff repair in order to provide a more meaningful comparison for other institutions, where indirect costs may be different.

In the literature, rigorous economic analysis of shoulder pathology is sparse. Kuye and colleagues¹² systematically reviewed economic evaluations in shoulder surgery for the period 1980–2010 and noted more than 50% of the papers were published between 2005 and 2010.¹² They also noted the poor quality of these studies and concluded more rigorous economic evaluations are needed to help justify the rising costs of shoulder-related treatments.

Several studies have directly evaluated costs associated with RCS. Cordasco and colleagues¹³ detailed the success of open rotator cuff repair as an outpatient procedure—noting its 43% cost savings ($4300 for outpatient vs $7500 for inpatient) and high patient satisfaction—using hospital charge data for operating room time, supplies, instruments, and postoperative slings. Churchill and Ghorai¹⁴ evaluated costs of mini-open and arthroscopic rotator cuff repairs in a statewide database and estimated the arthroscopic repair cost at $8985, compared with $7841 for the mini-open repair. They used reported hospital charge data, which were not itemized and did not include physician professional fees. Adla and colleagues,¹⁵ in a similar analysis of open versus arthroscopic cuff repair, estimated direct material costs of $1609.50 (arthroscopic) and $360.75 (open); these figures were converted from 2005 UK currency using the exchange rate cited in their paper. Salaries of surgeon, anesthesiologist, and other operating room personnel were said to be included in the operating room cost, but the authors’ paper did not include these data.

Two studies directly estimated the costs of arthroscopic rotator cuff repair. Hearnden and Tennent¹⁶ calculated the cost of RCS at their UK institution to be £2672, which included cost of operating room consumable materials, medication, and salaries of operating room personnel, including surgeon and anesthesiologist. Using online currency conversion from 2008 exchange rates and adjusting for inflation gave a corresponding US cost of $5449.63.¹⁷ Vitale and colleagues¹⁸ prospectively calculated costs of arthroscopic rotator cuff repair over a 1-year period using a cost-to-charge ratio from tabulated inpatient charges, procedure charges, and physician fees and payments abstracted from medical records, hospital billing, and administrative databases. Mean total cost for this cycle was $10,605.20, which included several costs (physical therapy, radiologist fees) not included in the present study. These studies, though more comprehensive than prior work, did not capture the entire cycle of surgical care.

Our study was designed to provide initial data on the direct costs of arthroscopic repair of the rotator cuff for the entire process cycle. Our overall cost estimate of $5904.21 differs significantly from prior work—not unexpected given the completely different cost methodology used.

Our study had several limitations. First, it was a single-surgeon evaluation, and a number of operating room variables (eg, use of adjunct instrumentation such as radiofrequency probes, differences in draping preferences) as well as surgeon volume in performing rotator cuff repairs might have substantially affected the reproducibility and generalizability of our data. Similarly, the large number of adjunctive procedures (eg, subacromial decompression, labral débridement) performed in conjunction with the rotator cuff repairs added operative time and therefore increased overall cost. Double-row repairs added operative time and increased the cost of consumable materials as well. Differences in surgeon preference for suture anchors may also be important, as anchors are a major cost driver and can vary significantly between vendors and institutions. Tear-related variables (eg, tear size, tear chronicity, degree of fatty cuff degeneration) were not controlled for and might have significantly affected operative time and associated cost. Resident involvement in the surgical procedure and anesthesia process in an academic setting prolongs surgical time and thus directly impacts costs.

In addition, we used the patient’s time in the operating room as a proxy for actual surgical time, as this was the only reliable and reproducible data point available in our electronic medical record. As such, an unquantifiable amount of surgeon time may have been overallocated to our cost estimate for time spent inducing anesthesia, positioning, helping take the patient off the operating table, and so on. However, as typical surgeon practice is to be involved in these tasks in the operating room, the possible overestimate of surgeon cost is likely minimal.

Our salary data for the TDABC algorithm were based on national averages for work hours and gross income for physicians and on hospital-based wage structure and may not be generalizable to other institutions. There may also be regional differences in work hours and salaries, which in turn would factor into a different per-minute cost for surgeon and anesthesiologist, depending on the exact geographic area where the surgery is performed. Costs may be higher at institutions that use certified nurse anesthetists rather than resident physicians because of the salary differences between these practitioners.

Moreover, the time that patients spend in the holding area—waiting to go into surgery and, after surgery, waiting for their ride home, for their prescriptions to be ready, and so forth—is an important variable to consider from a cost standpoint. However, as this time varied significantly and involved minimal contact with hospital personnel, we excluded its associated costs from our analysis. Similarly, and as already noted, hospital overhead and other indirect costs were excluded from analysis as well.

Conclusion

Using the TDABC algorithm, we found a direct economic cost of $5904.21 for RCS at our academic outpatient surgical center, with anchor cost the main cost driver. Judicious use of consumable resources is a key focus for cost containment in arthroscopic shoulder surgery, particularly with respect to implantable suture anchors. However, in the setting of more complex tears that require multiple anchors in a double-row repair construct, our pilot data may be useful to hospitals and surgery centers negotiating procedural reimbursement for the increased cost of complex repairs. Use of the TDABC algorithm for RCS and other procedures may also help in identifying opportunities to deliver more cost-effective health care.