“July Phenomenon” Revisited

Study Setting

TOH is a tertiary‐care teaching hospital with two inpatient campuses. The hospital operates within a publicly funded health care system, serves a population of approximately 1.5 million people in Ottawa and Eastern Ontario, treats all major trauma patients for the region, and provides most of the oncological care in the region.

TOH is the primary medical teaching hospital at the University of Ottawa. In 2010, there were 197 residents starting their first year of postgraduate training in one of 29 programs.

Inclusion Criteria

The study period extended from April 15, 2004 to December 31, 2008. We used this start time because our hospital switched to new coding systems for procedures and diagnoses in April 2002. Since these new coding systems contributed to our outcome statistic, we used a very long period (ie, two years) for coding patterns to stabilize to ensure that any changes seen were not a function of coding patterns. We ended our study in December 2008 because this was the last date of complete data at the time we started the analysis.

We included all medical, surgical, and obstetrical patients admitted to TOH during this time except those who were: younger than 15 years old; transferred to or from another acute care hospital; or obstetrical patients hospitalized for routine childbirth. These patients were excluded because they were not part of the multivariate model that we used to calculate risk of death in hospital (discussed below).24 These exclusions accounted for 25.4% of all admissions during the study period (36,820less than 15 years old; 12,931transferred to or from the hospital; and 44,220uncomplicated admission for childbirth).

All data used in this study came from The Ottawa Hospital Data Warehouse (TOHDW). This is a repository of clinical, laboratory, and administrative data originating from the hospital's major operational information systems. TOHDW contains information on patient demographics and diagnoses, as well as procedures and patient transfers between different units or hospital services during the admission.

Primary OutcomeRatio of Observed to Expected Number of Deaths per Week

For each study day, we measured the number of hospital deaths from the patient registration table in TOHDW. This statistic was collated for each week to ensure numeric stability, especially in our subgroup analyses.

We calculated the weekly expected number of hospital deaths using an extension of the Escobar model.24 The Escobar is a logistic regression model that estimated the probability of death in hospital that was derived and internally validated on almost 260,000 hospitalizations at 17 hospitals in the Kaiser Permanente Health Plan. It included six covariates that were measurable at admission including: patient age; patient sex; admission urgency (ie, elective or emergent) and service (ie, medical or surgical); admission diagnosis; severity of acute illness as measured by the Laboratory‐based Acute Physiology Score (LAPS); and chronic comorbidities as measured by the COmorbidity Point Score (COPS). Hospitalizations were grouped by admission diagnosis. The final model had excellent discrimination (c‐statistic 0.88) and calibration (P value of Hosmer Lemeshow statistic for entire cohort 0.66). This model was externally validated in our center with a c‐statistic of 0.901.25

We extended the Escobar model in several ways (Wong et al., Derivation and validation of a model to predict the daily risk of death in hospital, 2010, unpublished work). First, we modified it into a survival (rather than a logistic) model so it could estimate a daily probability of death in hospital. Second, we included the same covariates as Escobar except that we expressed LAPS as a time‐dependent covariate (meaning that the model accounted for changes in its value during the hospitalization). Finally, we included other time‐dependent covariates including: admission to intensive care unit; undergoing significant procedures; and awaiting long‐term care. This model had excellent discrimination (concordance probability of 0.895, 95% confidence interval [CI] 0.8890.902) and calibration.

We used this survival model to estimate the daily risk of death for all patients in the hospital each day. Summing these risks over hospital patients on each day returned the daily number of expected hospital deaths. This was collated per week.

The outcome statistic for this study was the ratio of the observed to expected weekly number of hospital deaths. Ratios exceeding 1 indicate that more deaths were observed than were expected (given the distribution of important covariates in those people during that week). This outcome statistic has several advantages. First, it accounts for the number of patients in the hospital each day. This is important because the number of hospital deaths will increase as the number of people in hospital increase. Second, it accounts for the severity of illness in each patient on each hospital day. This accounts for daily changes in risk of patient death, because calculation of the expected number of deaths per day was done using a multivariate survival model that included time‐dependent covariates. Therefore, each individual's predicted hazard of death (which was summed over the entire hospital to calculate the total expected number of deaths in hospital each day) took into account the latest values of these covariates. Previous analyses only accounted for risk of death at admission.

Expressing Physician Experience

The latent measure26 in all July Phenomenon studies is collective house‐staff physician experience. This is quantified by a surrogate date variable in which July 1the date that new house‐staff start their training in North Americarepresents minimal experience and June 30 represents maximal experience. We expressed collective physician experience on a scale from 0 (minimum experience) on July 1 to 1 (maximum experience) on June 30. A similar approach has been used previously13 and has advantages over the other methods used to capture collective house‐staff experience. In the stratified, incomplete approach,47, 911, 13, 1517 periods with inexperienced house‐staff (eg, July and August) are grouped together and compared to times with experienced house‐staff (eg, May and June), while ignoring all other data. The specification of cut‐points for this stratification is arbitrary and the method ignores large amounts of data. In the stratified, complete approach, periods with inexperienced house‐staff (eg, July and August) are grouped together and compared to all other times of the year.8, 12, 14, 1820, 22 This is potentially less biased because there are no lost data. However, the cut‐point for determining when house‐staff transition from inexperienced to experienced is arbitrary, and the model assumes that the transition is sudden. This is suboptimal because acquisition of experience is a gradual, constant process.

The pattern by which collective physician experience changes between July 1st and June 30th is unknown. We therefore expressed this evolution using five different patterns varying from a linear change to a natural logarithmic change (see Supporting Appendix B in the online version of this article).

Analysis

We first examined for autocorrelation in our outcome variable using Ljung‐Box statistics at lag 6 and 12 in PROC ARIMA (SAS 9.2, Cary, NC). If significant autocorrelation was absent in our data, linear regression modeling was used to associate the ratio of the observed to expected number of weekly deaths (the outcome variable) with the collective first year physician experience (the predictor variable). Time‐series methodology was to be used if significant autocorrelation was present.

In our baseline analysis, we included all hospitalizations together. In stratified analyses, we categorized hospitalizations by admission status (emergent vs elective) and admission service (medicine vs surgery).

Weekly Deaths: Observed, Expected, and Ratio

Figure 1A presents the observed weekly number of deaths during the study period. There was an average of 31 deaths per week (range 1551). Some large fluctuations in the weekly number of deaths were seen; in 2007, for example, the number of observed deaths went from 21 in week 13 up to 46 in week 15. However, no obvious seasonal trends in the observed weekly number of deaths were seen (Figure 1A, heavy line) nor were trends between years obvious.

Figure 1

Figure 1B presents the expected weekly number of deaths during the study period. The expected weekly number of deaths averaged 29.6 (range 22.238.7). The expected weekly number of deaths was notably less variable than the observed number of deaths. However, important variations in the expected number of deaths were seen; for example, in 2005, the expected number of deaths increased from 24.1 in week 41 to 29.6 in week 44. Again, we saw no obvious seasonal trends in the expected weekly number of deaths (Figure 1B, heavy line).

Figure 1C illustrates the ratio of observed to the expected weekly number of deaths. The average observed to expected ratio slightly exceeded unity (1.05) and ranged from 0.488 (week 24, in 2008) to 1.821 (week 51, in 2008). We saw no obvious seasonal trends in the ratio of the observed to expected number of weekly deaths. In addition, obvious trends in this ratio were absent over the study period.

Association Between House‐Staff Experience and Death in Hospital

We found no evidence of autocorrelation in the ratio of observed to expected weekly number of deaths. The ratio of observed to expected number of hospital deaths was not significantly associated with house‐staff physician experience (Table 2). This conclusion did not change regardless of which house‐staff physician experience pattern was used in the linear model (Table 2). In addition, our analysis found no significant association between physician experience and patient mortality when analyses were stratified by admission service or admission status (Table 2).

Study Setting

TOH is a tertiary‐care teaching hospital with two inpatient campuses. The hospital operates within a publicly funded health care system, serves a population of approximately 1.5 million people in Ottawa and Eastern Ontario, treats all major trauma patients for the region, and provides most of the oncological care in the region.

TOH is the primary medical teaching hospital at the University of Ottawa. In 2010, there were 197 residents starting their first year of postgraduate training in one of 29 programs.

Inclusion Criteria

The study period extended from April 15, 2004 to December 31, 2008. We used this start time because our hospital switched to new coding systems for procedures and diagnoses in April 2002. Since these new coding systems contributed to our outcome statistic, we used a very long period (ie, two years) for coding patterns to stabilize to ensure that any changes seen were not a function of coding patterns. We ended our study in December 2008 because this was the last date of complete data at the time we started the analysis.

We included all medical, surgical, and obstetrical patients admitted to TOH during this time except those who were: younger than 15 years old; transferred to or from another acute care hospital; or obstetrical patients hospitalized for routine childbirth. These patients were excluded because they were not part of the multivariate model that we used to calculate risk of death in hospital (discussed below).24 These exclusions accounted for 25.4% of all admissions during the study period (36,820less than 15 years old; 12,931transferred to or from the hospital; and 44,220uncomplicated admission for childbirth).

All data used in this study came from The Ottawa Hospital Data Warehouse (TOHDW). This is a repository of clinical, laboratory, and administrative data originating from the hospital's major operational information systems. TOHDW contains information on patient demographics and diagnoses, as well as procedures and patient transfers between different units or hospital services during the admission.

Primary OutcomeRatio of Observed to Expected Number of Deaths per Week

For each study day, we measured the number of hospital deaths from the patient registration table in TOHDW. This statistic was collated for each week to ensure numeric stability, especially in our subgroup analyses.

We calculated the weekly expected number of hospital deaths using an extension of the Escobar model.24 The Escobar is a logistic regression model that estimated the probability of death in hospital that was derived and internally validated on almost 260,000 hospitalizations at 17 hospitals in the Kaiser Permanente Health Plan. It included six covariates that were measurable at admission including: patient age; patient sex; admission urgency (ie, elective or emergent) and service (ie, medical or surgical); admission diagnosis; severity of acute illness as measured by the Laboratory‐based Acute Physiology Score (LAPS); and chronic comorbidities as measured by the COmorbidity Point Score (COPS). Hospitalizations were grouped by admission diagnosis. The final model had excellent discrimination (c‐statistic 0.88) and calibration (P value of Hosmer Lemeshow statistic for entire cohort 0.66). This model was externally validated in our center with a c‐statistic of 0.901.25

We extended the Escobar model in several ways (Wong et al., Derivation and validation of a model to predict the daily risk of death in hospital, 2010, unpublished work). First, we modified it into a survival (rather than a logistic) model so it could estimate a daily probability of death in hospital. Second, we included the same covariates as Escobar except that we expressed LAPS as a time‐dependent covariate (meaning that the model accounted for changes in its value during the hospitalization). Finally, we included other time‐dependent covariates including: admission to intensive care unit; undergoing significant procedures; and awaiting long‐term care. This model had excellent discrimination (concordance probability of 0.895, 95% confidence interval [CI] 0.8890.902) and calibration.

We used this survival model to estimate the daily risk of death for all patients in the hospital each day. Summing these risks over hospital patients on each day returned the daily number of expected hospital deaths. This was collated per week.

The outcome statistic for this study was the ratio of the observed to expected weekly number of hospital deaths. Ratios exceeding 1 indicate that more deaths were observed than were expected (given the distribution of important covariates in those people during that week). This outcome statistic has several advantages. First, it accounts for the number of patients in the hospital each day. This is important because the number of hospital deaths will increase as the number of people in hospital increase. Second, it accounts for the severity of illness in each patient on each hospital day. This accounts for daily changes in risk of patient death, because calculation of the expected number of deaths per day was done using a multivariate survival model that included time‐dependent covariates. Therefore, each individual's predicted hazard of death (which was summed over the entire hospital to calculate the total expected number of deaths in hospital each day) took into account the latest values of these covariates. Previous analyses only accounted for risk of death at admission.

Expressing Physician Experience

The latent measure26 in all July Phenomenon studies is collective house‐staff physician experience. This is quantified by a surrogate date variable in which July 1the date that new house‐staff start their training in North Americarepresents minimal experience and June 30 represents maximal experience. We expressed collective physician experience on a scale from 0 (minimum experience) on July 1 to 1 (maximum experience) on June 30. A similar approach has been used previously13 and has advantages over the other methods used to capture collective house‐staff experience. In the stratified, incomplete approach,47, 911, 13, 1517 periods with inexperienced house‐staff (eg, July and August) are grouped together and compared to times with experienced house‐staff (eg, May and June), while ignoring all other data. The specification of cut‐points for this stratification is arbitrary and the method ignores large amounts of data. In the stratified, complete approach, periods with inexperienced house‐staff (eg, July and August) are grouped together and compared to all other times of the year.8, 12, 14, 1820, 22 This is potentially less biased because there are no lost data. However, the cut‐point for determining when house‐staff transition from inexperienced to experienced is arbitrary, and the model assumes that the transition is sudden. This is suboptimal because acquisition of experience is a gradual, constant process.

The pattern by which collective physician experience changes between July 1st and June 30th is unknown. We therefore expressed this evolution using five different patterns varying from a linear change to a natural logarithmic change (see Supporting Appendix B in the online version of this article).

Analysis

We first examined for autocorrelation in our outcome variable using Ljung‐Box statistics at lag 6 and 12 in PROC ARIMA (SAS 9.2, Cary, NC). If significant autocorrelation was absent in our data, linear regression modeling was used to associate the ratio of the observed to expected number of weekly deaths (the outcome variable) with the collective first year physician experience (the predictor variable). Time‐series methodology was to be used if significant autocorrelation was present.

In our baseline analysis, we included all hospitalizations together. In stratified analyses, we categorized hospitalizations by admission status (emergent vs elective) and admission service (medicine vs surgery).

Weekly Deaths: Observed, Expected, and Ratio

Figure 1A presents the observed weekly number of deaths during the study period. There was an average of 31 deaths per week (range 1551). Some large fluctuations in the weekly number of deaths were seen; in 2007, for example, the number of observed deaths went from 21 in week 13 up to 46 in week 15. However, no obvious seasonal trends in the observed weekly number of deaths were seen (Figure 1A, heavy line) nor were trends between years obvious.

Figure 1

Figure 1B presents the expected weekly number of deaths during the study period. The expected weekly number of deaths averaged 29.6 (range 22.238.7). The expected weekly number of deaths was notably less variable than the observed number of deaths. However, important variations in the expected number of deaths were seen; for example, in 2005, the expected number of deaths increased from 24.1 in week 41 to 29.6 in week 44. Again, we saw no obvious seasonal trends in the expected weekly number of deaths (Figure 1B, heavy line).

Figure 1C illustrates the ratio of observed to the expected weekly number of deaths. The average observed to expected ratio slightly exceeded unity (1.05) and ranged from 0.488 (week 24, in 2008) to 1.821 (week 51, in 2008). We saw no obvious seasonal trends in the ratio of the observed to expected number of weekly deaths. In addition, obvious trends in this ratio were absent over the study period.

Association Between House‐Staff Experience and Death in Hospital

We found no evidence of autocorrelation in the ratio of observed to expected weekly number of deaths. The ratio of observed to expected number of hospital deaths was not significantly associated with house‐staff physician experience (Table 2). This conclusion did not change regardless of which house‐staff physician experience pattern was used in the linear model (Table 2). In addition, our analysis found no significant association between physician experience and patient mortality when analyses were stratified by admission service or admission status (Table 2).

Study Setting

TOH is a tertiary‐care teaching hospital with two inpatient campuses. The hospital operates within a publicly funded health care system, serves a population of approximately 1.5 million people in Ottawa and Eastern Ontario, treats all major trauma patients for the region, and provides most of the oncological care in the region.

TOH is the primary medical teaching hospital at the University of Ottawa. In 2010, there were 197 residents starting their first year of postgraduate training in one of 29 programs.

Inclusion Criteria

The study period extended from April 15, 2004 to December 31, 2008. We used this start time because our hospital switched to new coding systems for procedures and diagnoses in April 2002. Since these new coding systems contributed to our outcome statistic, we used a very long period (ie, two years) for coding patterns to stabilize to ensure that any changes seen were not a function of coding patterns. We ended our study in December 2008 because this was the last date of complete data at the time we started the analysis.

We included all medical, surgical, and obstetrical patients admitted to TOH during this time except those who were: younger than 15 years old; transferred to or from another acute care hospital; or obstetrical patients hospitalized for routine childbirth. These patients were excluded because they were not part of the multivariate model that we used to calculate risk of death in hospital (discussed below).24 These exclusions accounted for 25.4% of all admissions during the study period (36,820less than 15 years old; 12,931transferred to or from the hospital; and 44,220uncomplicated admission for childbirth).

All data used in this study came from The Ottawa Hospital Data Warehouse (TOHDW). This is a repository of clinical, laboratory, and administrative data originating from the hospital's major operational information systems. TOHDW contains information on patient demographics and diagnoses, as well as procedures and patient transfers between different units or hospital services during the admission.

Primary OutcomeRatio of Observed to Expected Number of Deaths per Week

For each study day, we measured the number of hospital deaths from the patient registration table in TOHDW. This statistic was collated for each week to ensure numeric stability, especially in our subgroup analyses.

We calculated the weekly expected number of hospital deaths using an extension of the Escobar model.24 The Escobar is a logistic regression model that estimated the probability of death in hospital that was derived and internally validated on almost 260,000 hospitalizations at 17 hospitals in the Kaiser Permanente Health Plan. It included six covariates that were measurable at admission including: patient age; patient sex; admission urgency (ie, elective or emergent) and service (ie, medical or surgical); admission diagnosis; severity of acute illness as measured by the Laboratory‐based Acute Physiology Score (LAPS); and chronic comorbidities as measured by the COmorbidity Point Score (COPS). Hospitalizations were grouped by admission diagnosis. The final model had excellent discrimination (c‐statistic 0.88) and calibration (P value of Hosmer Lemeshow statistic for entire cohort 0.66). This model was externally validated in our center with a c‐statistic of 0.901.25

We extended the Escobar model in several ways (Wong et al., Derivation and validation of a model to predict the daily risk of death in hospital, 2010, unpublished work). First, we modified it into a survival (rather than a logistic) model so it could estimate a daily probability of death in hospital. Second, we included the same covariates as Escobar except that we expressed LAPS as a time‐dependent covariate (meaning that the model accounted for changes in its value during the hospitalization). Finally, we included other time‐dependent covariates including: admission to intensive care unit; undergoing significant procedures; and awaiting long‐term care. This model had excellent discrimination (concordance probability of 0.895, 95% confidence interval [CI] 0.8890.902) and calibration.

We used this survival model to estimate the daily risk of death for all patients in the hospital each day. Summing these risks over hospital patients on each day returned the daily number of expected hospital deaths. This was collated per week.

The outcome statistic for this study was the ratio of the observed to expected weekly number of hospital deaths. Ratios exceeding 1 indicate that more deaths were observed than were expected (given the distribution of important covariates in those people during that week). This outcome statistic has several advantages. First, it accounts for the number of patients in the hospital each day. This is important because the number of hospital deaths will increase as the number of people in hospital increase. Second, it accounts for the severity of illness in each patient on each hospital day. This accounts for daily changes in risk of patient death, because calculation of the expected number of deaths per day was done using a multivariate survival model that included time‐dependent covariates. Therefore, each individual's predicted hazard of death (which was summed over the entire hospital to calculate the total expected number of deaths in hospital each day) took into account the latest values of these covariates. Previous analyses only accounted for risk of death at admission.

Expressing Physician Experience

The latent measure26 in all July Phenomenon studies is collective house‐staff physician experience. This is quantified by a surrogate date variable in which July 1the date that new house‐staff start their training in North Americarepresents minimal experience and June 30 represents maximal experience. We expressed collective physician experience on a scale from 0 (minimum experience) on July 1 to 1 (maximum experience) on June 30. A similar approach has been used previously13 and has advantages over the other methods used to capture collective house‐staff experience. In the stratified, incomplete approach,47, 911, 13, 1517 periods with inexperienced house‐staff (eg, July and August) are grouped together and compared to times with experienced house‐staff (eg, May and June), while ignoring all other data. The specification of cut‐points for this stratification is arbitrary and the method ignores large amounts of data. In the stratified, complete approach, periods with inexperienced house‐staff (eg, July and August) are grouped together and compared to all other times of the year.8, 12, 14, 1820, 22 This is potentially less biased because there are no lost data. However, the cut‐point for determining when house‐staff transition from inexperienced to experienced is arbitrary, and the model assumes that the transition is sudden. This is suboptimal because acquisition of experience is a gradual, constant process.

The pattern by which collective physician experience changes between July 1st and June 30th is unknown. We therefore expressed this evolution using five different patterns varying from a linear change to a natural logarithmic change (see Supporting Appendix B in the online version of this article).

Analysis

We first examined for autocorrelation in our outcome variable using Ljung‐Box statistics at lag 6 and 12 in PROC ARIMA (SAS 9.2, Cary, NC). If significant autocorrelation was absent in our data, linear regression modeling was used to associate the ratio of the observed to expected number of weekly deaths (the outcome variable) with the collective first year physician experience (the predictor variable). Time‐series methodology was to be used if significant autocorrelation was present.

In our baseline analysis, we included all hospitalizations together. In stratified analyses, we categorized hospitalizations by admission status (emergent vs elective) and admission service (medicine vs surgery).

Weekly Deaths: Observed, Expected, and Ratio

Figure 1A presents the observed weekly number of deaths during the study period. There was an average of 31 deaths per week (range 1551). Some large fluctuations in the weekly number of deaths were seen; in 2007, for example, the number of observed deaths went from 21 in week 13 up to 46 in week 15. However, no obvious seasonal trends in the observed weekly number of deaths were seen (Figure 1A, heavy line) nor were trends between years obvious.

Figure 1

Figure 1B presents the expected weekly number of deaths during the study period. The expected weekly number of deaths averaged 29.6 (range 22.238.7). The expected weekly number of deaths was notably less variable than the observed number of deaths. However, important variations in the expected number of deaths were seen; for example, in 2005, the expected number of deaths increased from 24.1 in week 41 to 29.6 in week 44. Again, we saw no obvious seasonal trends in the expected weekly number of deaths (Figure 1B, heavy line).

Figure 1C illustrates the ratio of observed to the expected weekly number of deaths. The average observed to expected ratio slightly exceeded unity (1.05) and ranged from 0.488 (week 24, in 2008) to 1.821 (week 51, in 2008). We saw no obvious seasonal trends in the ratio of the observed to expected number of weekly deaths. In addition, obvious trends in this ratio were absent over the study period.

Association Between House‐Staff Experience and Death in Hospital

We found no evidence of autocorrelation in the ratio of observed to expected weekly number of deaths. The ratio of observed to expected number of hospital deaths was not significantly associated with house‐staff physician experience (Table 2). This conclusion did not change regardless of which house‐staff physician experience pattern was used in the linear model (Table 2). In addition, our analysis found no significant association between physician experience and patient mortality when analyses were stratified by admission service or admission status (Table 2).