Affiliations of authors: C. J. Newschaffer, Department of Epidemiology, The Johns Hopkins School of Hygiene and Public Health, Baltimore, MD, and Department of Community Health, Saint Louis University School of Public Health, St. Louis, MO; K. Otani, Department of Community Health, Saint Louis University School of Public Health; M. K. McDonald, L. T. Penberthy, Department of Medicine, Division of Quality Health Care, Medical College of Virginia, Virginia Commonwealth University, Richmond.
Correspondence to: Craig J. Newschaffer, Ph.D., Department of Epidemiology, The Johns Hopkins School of Hygiene and Public Health, 615 N. Wolfe St., Baltimore, MD 21205 (e-mail: cnewscha{at}jhsph.edu).
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ABSTRACT |
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INTRODUCTION |
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Attribution of the underlying cause of death in older individuals with multiple chronic disease can be a difficult process (5), and only limited work has been done describing clinical and demographic factors associated with the ascribed cause of death among prostate cancer patients. Associations between age, race, stage, initial treatment, and comorbid cardiovascular disease with prostate cancer underlying cause of death have been recently reported among decedents in a prostate cancer patient cohort (6). However, equally important to understanding influences on cause-of-death reporting in prostate cancer patients is an examination of nonprostate cancer causes of death. Furthermore, examination of the distribution of causes of death among prostate cancer patients will be more informative when compared with the distribution of causes of death among men without the disease. Here, we explore the underlying causes of death listed for decedents in a large, population-based cohort of prostate cancer case subjects and compare their patterns of nonprostate cancer deaths to those among decedents from a cohort of men without prostate cancer.
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SUBJECTS AND METHODS |
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The population under study consisted of two groups of decedents. The first was 1207 decedents from a cohort of 1996 prostate cancer patients residing in Virginia who were diagnosed from 1987 through 1989. Decedents from this cohort were ascertained through record linkage (described below under "Vital Status Follow-up") completed in 1997, ascertaining deaths through the end of 1995. All patients were aged 67 years or more at diagnosis. Data on case subjects were drawn from Medicare and Virginia Cancer Registry (VCR) databases. As discussed previously (7), neither data source has perfect case ascertainment sensitivity and specificity, and the two sources likely include complementary data. Therefore, subjects with cancer were followed only if there was successful linkage between Medicare and VCR identifying information and if both sources showed an incident prostate cancer diagnosis within 6 months. Linkage was accomplished with the use of a three-step algorithm matching Social Security number (SSN), first name, last name, sex, and date of birth. The definition of a first-incident prostate cancer based on Medicare inpatient claims data is shown in Appendix I. The age threshold of 67 years was adopted to ensure that all men had at least 2 years of Medicare eligibility prior to diagnosis to allow for review of comorbidity.
The second group of 2906 decedents was drawn from a nonprostate cancer cohort (n = 6586) comprising male Medicare beneficiaries aged 67 years or more living in Virginia who were hospitalized for benign prostatic hyperplasia (BPH) from 1987 through 1989 with no history of prostate cancer. Medicare data were used to identify these men (criteria described in Appendix I), and lack of prostate cancer history was determined via review of 2 years' previous Medicare inpatient claims and a crosscheck with linked VCR records. Since Medicare hospital inpatient diagnostic codes were the only available source of comorbidity data, we needed to select a nonprostate cancer group with at least one hospitalization so that there was equivalent opportunity to report comorbidity. A hospitalization for BPH was selected because the disease has minimal direct effect on survival but typically confers morbidity that, during the study period, commonly led to inpatient interventions (8,9). Pathologic analysis of tissue excised during partial resection for suspected BPH could lead to a diagnosis of prostate cancer; however, these patients would have prostate cancer coded in the first diagnostic position on the Medicare claim and, consequently, would be included in the prostate cancer, not the BPH, group. Record linkage to identify decedents in this cohort followed the same procedure as that for the prostate cancer cohort. The protocol for linkage and analysis of study data was approved by human subjects committees at Saint Louis University (St. Louis, MO) and the Centers for Disease Control and Prevention (Atlanta, GA).
Vital Status Follow-up
Vital status was ascertained in hierarchic fashion by use of Virginia Department of Health (VDH) Vital Statistics, National Death Index Plus (NDI-Plus), and, when necessary, death certificates supplied from states other than Virginia. First, VDH vital statistics files were matched to identifiers of subjects in the cancer and nonprostate cancer cohorts. When SSNs were available, matches had to agree exactly on the SSN and, allowing for minor misspellings and suffix differences, the name. In cases where no SN was available, matches were based on exact name and date-of-birth agreement. VDH vital statistics files contain information on the date and cause of death. The underlying cause of death is included as coded by state nosologists completing death certificate review for submission to the National Center for Vital Statistics. Fields are also available for up to four additional causes of death.
To identify members of each cohort dying outside the state of Virginia, identifier information for all subjects not matching VDH vital statistics files was submitted for NDI-Plus database matching. The criteria for accepting matched deaths from NDI-Plus are described in Appendix Table II. The NDI-Plus database contains information on date and, for decedents in participating states, cause of death. An additional 54 decedents were identified. For those identified from states not participating in the cause-of-death reporting component of NDI-Plus (19 of the 54 subjects), we obtained death certificates directly from state health departments. By special arrangement, the code combinations from these death certificates were reviewed by state of Virginia nosologists who applied their standard algorithm for vital statistics reporting to assign an underlying cause of death (10).
Underlying causes of death were categorized as prostate cancer, heart disease, other cancer, cerebrovascular disease, chronic obstructive pulmonary disease (COPD), pneumonia, diabetes, nephritis, septicemia, Alzheimer's disease, hypertension, and other. Throughout this article, the category "other cancer" refers to cancer at any site other than the prostate. The specific conditions used, other than prostate cancer, are the 10 most common underlying causes of death in white and African-American men over age 65 years in 1995 (11).
Other Study Variables
The Medicare files were the source of information on patient age at diagnosis and race. In these files, race was coded only as white, black, and other. For the cancer cohort, the VCR provided information on summary tumor stage (local, regional, or distant) at diagnosis. Unfortunately, during the study period, only 17% of the men diagnosed with prostate cancer who were reported to the VCR had more detailed American Joint Committee on Cancer staging data available (12).
Both the VCR and Medicare claims include data on the initial course of treatment of cancer patients. The VCR considers the first four postdiagnostic months as the initial treatment period. Therefore, we combined VCR initial treatment data with inpatient diagnostic and procedure codes from the index prostate cancer Medicare claim and any from the subsequent 4 months. We categorized initial treatment in several different ways. First, we used a five-level classification: no treatment, surgery, radiation therapy, combination therapy (either surgery or radiation therapy in combination with another treatment modality), and other (includes orchiectomy and/or hormonal treatment without surgery or radiation therapy). For some analyses, the categories of surgery, radiation therapy, and combination therapy as defined above were combined to form a single aggressive initial treatment category. Finally, we also subdivided the combination therapy category by whether or not surgery or radiation therapy was involved and combined these subjects with the respective surgery- or radiation therapy-only groups. Under this classification, 14 patients initially classified in the combination-therapy group had indications of both surgery and radiation therapy and were excluded.
Medicare claims also provided data on comorbidity. Each subject's index hospital claim as well as any other hospital claim in the previous 2 years provided up to five International Classification of Diseases, 9th Revision, Clinical Modification (ICD-9-CM), diagnosis codes (13). Information on individual comorbid conditions was aggregated with the use of the Charlson Comorbidity Index (CCI) (14). Previous work (15) has found the CCI to be an acceptable means of measuring aggregate comorbidity in prostate cancer cohorts. The original CCI was designed for use with medical record data, but it has been adapted to ICD-9-CM codes. We used the DartmouthManitoba ICD-9-CM algorithm (16), omitting the indicator for prostate cancer and cancer metastases. The CCI measured through administrative claims previously has been associated with mortality (1720) and a variety of other health outcomes (18,20). CCI scores were initially categorized as 0, 1, 2, and 3 or more; however, because of the low prevalence of CCI values greater than 1, we dichotomized CCI as present (score 1) versus absent in many analyses.
As a community-level measure of socioeconomic status, data on last grade of school completed by residents in subjects' ZIP Codes were obtained from 1990 census data (1990 Census Bureau Summary Tape 3b). The category of educational level matched to each subject was based on the proportion of individuals aged 65 years in the subject's race group with less than a 9th grade education residing in their ZIP Code. Categories were 0%5%, 6%15%, 16%20%, and 21% or more.
Statistical Analyses
The frequency distributions of the underlying causes of death in the prostate cancer and nonprostate cancer decedents were compared. This proportion of deaths for any cause is also known as the proportionate mortality rate (21). For each cause, a two-tailed test for the difference between independent binomial proportions (22) was used to compare proportionate mortality rates across prostate cancer and nonprostate cancer cohort decedents. Among prostate cancer decedents only, logistic regression models estimated odds ratios (ORs) for prostate cancer underlying cause of death associated with demographic, stage, treatment, and comorbidity variables. Time from prostate cancer diagnosis until death was included in all models as a five-level system of dummy variables (<12, 1324, 2548, 4972, and 73 months).
Next, the proportionate mortality rates for the leading underlying causes of death in the subgroup of prostate cancer cohort decedents dying of causes other than prostate cancer were compared with those of the nonprostate cancer cohort with accompanying tests of differences in proportions. The nonprostate cancer population is made up of BPH patients who likely share risk factors (e.g., endogenous hormone levels). Consequently, background mortality should be similar and, under the assumption that deaths with prostate cancer listed as the underlying cause completely captured mortality attributable to the disease, the distribution of other causes of death in prostate cancer patients is expected to be similar to that of the nonprostate cancer population.
Proportionate mortality rate comparisons were also made stratifying both the decedents from the prostate cancer cohort dying of nonprostate cancer causes and the nonprostate cancer cohort decedents by age, race, educational level, comorbidity, and, in the former group of decedents only, tumor stage and initial treatment. Multivariable logistic regression models were constructed to conduct adjusted comparisons of the odds of particular causes of death that differed substantively (absolute differences in proportionate mortality rate >5%) and statistically significantly (i.e., two-sided P<.05) in unadjusted comparisons across these groups.
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RESULTS |
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Next, analyses were conducted comparing proportional mortality in the subgroup of prostate cancer cohort decedents with underlying causes other than prostate cancer to proportionate mortality rates from the nonprostate cancer cohort. Fig. 2 shows that proportionate mortality (for all of the leading causes) across these groups are very similar, consistent with the hypothesis that prostate cancer patients who have died of causes other than prostate cancer have a cause-of-death distribution similar to that of the general population. When comparisons were stratified by age, race, comorbidity, ecologic education, and, for the prostate cancer group, stage, this similarity persisted (results not shown). However, when this prostate cancer group was stratified by initial treatment, discrepancies emerged.
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DISCUSSION |
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In a previous study of prostate cancer proportionate mortality, Satariano et al. (6) found cardiovascular comorbidity, but not other comorbidities, to be associated with decreased odds of a prostate cancer underlying cause of death. On the basis of this finding, these authors speculated that cardiovascular comorbidity may have special prognostic significance for prostate cancer patients. We, too, found cardiovascular comorbidity to reduce the odds of a prostate cancer cause of death, yet we believe that the negative association between cardiovascular comorbidity and prostate cancer proportionate mortality could occur even if there was no special prognostic significance of cardiovascular disease in prostate cancer patients. Like any competing risk, deaths due to the independent mortality risk posed by cardiovascular disease would dilute the contribution of prostate cancer deaths to the total decedent pool and thereby lower prostate cancer proportionate mortality. Because cardiovascular disease is the leading competing risk, it is not surprising that the association between cardiovascular disease and prostate cancer proportionate mortality is large and statistically significant; however, this association does not mean cardiovascular disease is any more of a threat in prostate cancer patients than in other men.
Satariano et al. (6) also observed no difference in prostate cancer proportionate mortality in patients with noncardiovascular comorbidity and in patients without comorbidity. However, we saw a similar magnitude-negative association with prostate cancer underlying cause for other comorbid diseases as for cardiovascular disease. Differences in comorbidity data sources or coding algorithms could contribute to this heterogeneity across studies.
We also performed a series of analyses focusing on the group of prostate cancer cohort decedents who did not have prostate cancer listed as their underlying cause of death. The rationale for this analysis is that, in the absence of major difficulties in the attribution of underlying cause of death, these decedents should have the same distribution of underlying causes of death as a comparison cohort of elderly men without prostate cancer. For the prostate cancer cohort overall, we found this to be the case.
This result is generally consistent with other studies that have examined the quality of vital statistics reporting of prostate cancer cause of death. Researchers from the National Cancer Institute (NCI), Bethesda, MD, have periodically examined underlying cancer cause of death data in large cohorts of cancer patients (2325). Two of the NCI studies (23,24) focused on the proportion of cancer patient decedents with an underlying cancer cause who had agreement between underlying cause and primary incident cancer sites (deemed the "detection rate"). For incident prostate cancer case subjects, this rate was 93.5%, one of the higher detection rates reported among the major cancer sites. If we estimate the detection rate from our data, we find a much lower rate of 77%; however, our elderly cohort excludes the 20% of prostate cancer patients diagnosed under the age of 65 years (26,27). This excluded group has a higher likelihood of dying of their disease; consequently, excluding them drives down this detection rate.
The other NCI investigation (25) compared population cancer-specific mortality rates derived from vital statistics (deaths with a cancer underlying cause over total population at risk, not just cancer patients) with life table-derived, cancer-specific mortality rates based on Surveillance, Epidemiology, and End Results (SEER)1-based incidence and relative survival rate estimates. Relative survival rates capture cancer-related mortality but do not rely on attribution of specific causes of death. For prostate cancer, vital statistics-based and life table-derived population mortality was similar at 23.1 and 24.7 per 100000, respectively. Note that, while these rates were estimated for the general population, if restricted to those with prostate cancer, the percentage difference between the rates would not change.
However, these studies did not examine particular subgroups of cancer patients. It is conceivable that there were problems in cause of death reporting within subgroups of prostate cancer patients and that these may have offset each other when all of the patients were examined together. In fact, this is what was suggested from our data with respect to prostate cancer subgroups defined by initial treatment. Decedents treated initially with watchful waiting were less likely to have a different cancer as the underlying cause of death than comparison cohort decedents, while decedents aggressively treated were more likely to have an other cancer as an underlying cause. This intriguing pattern persisted after adjustment for known covariates. We were, however, unable to create prostate cancer subgroups defined simultaneously by stage and initial treatment because of the small size of a number of the resulting strata. Separate analyses that compared local-, regional-, and distant-stage prostate cancer groups with the nonprostate cancer cohort showed no differences in other cancer proportionate mortality. This result indicates that the findings with respect to initial treatment were not likely confounded by stage of disease at diagnosis.
One possible explanation for the differing distribution of other cancer deaths in initial treatment groups is information bias; i.e., assignment of the underlying cause is dependent on knowledge of treatment approach. Health-care providers completing death certificates for prostate cancer patients known to have received aggressive treatments may have been less likely to attribute deaths to the prostate tumor because of their beliefs regarding the effectiveness of aggressive treatment. Such a practice might result in a higher frequency of other cancer codes because patients with cachexia and other signs of progressive neoplasia, if not assigned to prostate cancer, would likely be assigned to another type of cancer. Conversely, among patients known to have been followed with observation only, there may have been an increased tendency on the part of health-care providers to code deaths consistent with sequelae of late-stage cancer to the prostate tumor because of the lack of definitive initial therapy. There was no particular site disproportionately represented among the other cancer decedents in either treatment group.
Because our definition of aggressive treatment included radiation therapy and Medicare claim codes for radiation therapy are not site specific, we considered the possibility that case subjects designated as aggressively treated might include a greater proportion of individuals with coprimary cancers. If this were the case, the higher proportion of decedents with other cancer underlying causes seen in the aggressively treated group could be due solely to this enrichment with individuals having coprimary cancer. However, this does not appear likely because the prostate cancer cohort decedents with other cancer underlying cause of death who were aggressively treated actually had a lower proportion with indications of coprimary cancer than those with no initial treatment or other initial treatment. In addition, the proportion of other cancer deaths was actually higher among aggressively treated prostate cancer case subjects who had prostate surgery than among those who had radiation therapy.
The few available direct comparisons of underlying cause of death reported on vital statistics to a "gold standard" cause of death determination in prostate cancer case series do not rule out information bias. Unfortunately, studies that have used autopsy data as the gold standard (2830) have generally been comprehensive investigations of all major causes of death and consequently have not stratified by cancer site or attempted to identify prevalent cancer case subjects among decedents dying of other causes. Recently, however, there have been studies focusing on prostate cancer patients that used medical records reviews as a gold standard [(6); Funkhouser E: manuscript in preparation]. Percent agreement between the death certificate and the record review ranged from approximately 80% to 90%. The possibility that underlying cause of death might be misreported in as many as one fifth of all prostate cancer decedents does not appear to be unreasonable. Our findings suggest that a focus of future direct validation studies might be on other cancer cause of death and that special attention should be paid to stratifying by initial treatment.
There are, of course, a number of limitations to the data and methods that we used here. Although great care was taken in establishing linkages, there may still be errors in matching cancer registry, Medicare, and vital statistics data. In addition, while the Medicare and cancer registry data sources do not impose severe restrictions on the study population, the VCR was not yet fully population based (it included hospitals comprising 85% of the beds in the state at that time), and the Medicare data excluded a small fraction of patients with incomplete Medicare enrollment or a Health Maintenance Organization affiliation during the study period. Also, bear in mind that available information on comorbidity was imperfect, coming from administrative data,and disease severity and initial treatment data were restricted to what was available from the cancer registry. In interpreting our findings, we were careful to remember that factors associated with increased proportionate mortality are not necessarily associated with increased mortality risk (11); we urge readers to likewise keep this in mind.
Underlying cause of death data as reported in vital statistics is relied on heavily in population-based surveillance and observational epidemiologic studies. Most cancer registries, including SEER, employ vital records linkages as their principal means of incorporating cause of death information into their databases. Several recent prostate cancer studies, including several focusing on treatment, have used prostate cancer-specific mortality as ascertained from vital statistics, either directly or via a cancer registry, as a primary end point. This study, however, points out the possibility of an important limitation in vital statistics data, i.e., nondifferential misclassification of cause of death according to initial treatment status. We realize that our data analysis approach involved a large number of proportionate mortality rate comparisons, with only one major difference emerging. However, we agree with the perspective on multiple comparisons advanced by a number of authors (3133) where, in addition to bearing in mind the increased likelihood of such a finding being attributable to chance, the reader is also urged to consider carefully the plausibility of the hypothesis. The hypothesis of information bias that we propose seems sufficiently reasonable to warrant further consideration. Cause of death across initial treatment groups should be investigated thoroughly in direct comparisons of vital statistics-based cause of death information to medical records review and autopsy.
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Appendix I. |
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Prostate Cancer
Either: 1) first position malignant neoplasm of the prostate code (185) or 2) second through fifth position 185 with either: a. a prostatectomy procedure code (60.3, 60.4, 60.5, or 60.6); b. a transurethral prostatectomy (TURP) procedure code (60.2); or c. a biopsy procedure code (60.11 or 60.12) in any procedure code field.
The additional presence of a diagnostic code indicating personal history of prostate cancer (V10.46) would result in exclusion.
The presence of a diagnostic code of 185 or V10.46 on any inpatient claim 2 years prior to a claim meeting the above criteria would also result in an exclusion.
Benign Prostatic Hyperplasia
Either: 1) first position benign prostatic hyperplasia (600) or 2) second through fifth position 600 with either: a. a bladder neck obstruction diagnostic code (596.0); b. a retention of urine diagnostic code (788.2); c. a urinary tract infection, site not specified, diagnostic code (599.0) in the first diagnostic code field.
The presence of a diagnostic code of 600 on any inpatient claim 2 years prior to a claim meeting the above criteria would also result in an exclusion.
Note: Any men meeting the above criteria for either prostate cancer or BPH who were not continuously enrolled in Medicare or who belonged to a Medicare Health Maintenance Organization in the year of diagnosis or any of the prior 2 years are excluded, since all of their utilization data are not captured in the Medicare files.
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Appendix II. |
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NOTES |
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Supported by the Centers for Disease Control and Prevention under the Prevention Research Centers Cooperative Agreement No. U48/CCU710806-03.
We thank Ron Hyman and the nosologists of the Virginia Department of Health for their review of causes of death listed on non-Virginia death certificates.
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