Abstract
Background
Randomized clinical trials have shown the efficacy of adjuvant chemotherapy in treating node-positive operable breast cancer in women aged ≤69 years, but the benefit of chemotherapy in women aged ≥70 is questionable. This study was to examine if adjuvant chemotherapy is effective for these women with breast cancer.
Methods
We studied a cohort of 5464 women diagnosed with node-positive operable breast cancer at age ≥65 in 1992 through 1996 with last follow-up of December 31, 1999 in five states and six metropolitan areas. Hazard ratio (HR) for all-cause mortality was used for survival analysis with adjustment for patient and tumor characteristics; propensity analysis was used to control for observed factors; and sensitivity analysis was used to estimate potential effects of unmeasured confounders.
Results
After adjusting for propensity to receive chemotherapy, the chemotherapy-treated and untreated groups were not statistically significantly different for covariates except for age and hormone receptor status. Mortality was significantly reduced in women aged 65–69 who received adjuvant chemotherapy compared to those who did not, after adjusting for patient and tumor characteristics (HR = 0.70, 95% confidence interval [CI], 0.57–0.88) or after adjusting for propensity scores (HR = 0.76, 95% CI, 0.62–0.94). HR did not significantly differ between the treated and untreated women aged ≥70 (HR = 0.96, 95% CI = 0.83–1.09, and HR = 0.99, 95% CI, 0.87–1.14). These results were relatively insensitive to changes in unmeasured confounders.
Conclusions
Adjuvant chemotherapy is associated with improved survival in women with node-positive operable breast cancer aged 65–69 living in the community, but not in women aged ≥70. These findings are consistent with those found in randomized controlled trials.
Over the past three decades, numerous randomized controlled trials have demonstrated that chemotherapy improves both recurrence-free survival and overall survival for women with early-stage breast cancer (1–5). However, the efficacy of adjuvant chemotherapy has been shown to decrease with age. For example, the pooled analysis of 47 randomized trials by the Early Breast Cancer Trialists' Collaborative Group showed that chemotherapy provided a 27% proportional reduction in 10-year mortality for women with early-stage operable breast cancer aged <50, 14% for women aged 50–59, and 8% for women aged 60–69 (3). This translates to an absolute benefit in 10-year survival of 7%–11% for those aged <50 and 2%–3% for those aged 50–69 years (3). This meta-analysis found no benefit of chemotherapy in women aged ≥70. Because of these findings, the recommendations for chemotherapy use in women aged ≥70 are not clear cut, with most authorities stressing the need for decision making based on the particular condition of the individual patient (6–10).
Randomized controlled clinical trials are considered to be the gold standard for determining the efficacy of a therapy. However, participants in clinical trials do not represent a cross-section of patients in the community (11–14). Population-based observational studies can therefore be useful adjuncts to randomized controlled trials in determining whether efficacy under controlled conditions in special centers translates into real-world effectiveness in routine practice (11–13). Furthermore, in some circumstances where the randomized trials are not possible due to ethical or logistic issues, well-conducted observational studies could potentially function effectively in determining drug efficacy (11).
We used the National Cancer Institute's Surveillance, Epidemiology, and End Results (SEER) cancer registry data, accounting for 14% of the total U.S. population, linked to Medicare data, to examine the effectiveness of adjuvant chemotherapy in older women with breast cancer. We hypothesized that chemotherapy is effective in treating node-positive operable breast cancer in women aged 65–69 years at diagnosis, consistent with what has been shown in clinical trials. Furthermore, given a large number of patients aged ≥70 in our study database, we wanted to assess whether adjuvant chemotherapy is effective in this population.
Methods
Data Sources
The SEER program, supported by the National Cancer Institute, includes population-based tumor registries in 11 selected geographic areas (15,16): the metropolitan areas of San Francisco/Oakland, Detroit, Atlanta, and Seattle; Los Angeles county; the San Jose–Monterey area; and the states of Connecticut, Iowa, New Mexico, Utah, and Hawaii.
The Medicare Program covers hospital, physician, and outpatient medical services for more than 97% of persons aged ≥65 (17,18). These data were available for all beneficiaries starting in 1991, and their Medicare claims are available through 1999. Cases reported by the SEER registries from 1992 to 1996 were matched against the Medicare master enrollment file. Medicare eligibility could be identified for 94% of the persons aged ≥65 appearing in the SEER records. The method of linking these data has been described elsewhere (17,19). The Institutional Review Boards of the University of Texas Health Science Center at Houston and University of Texas Medical Branch at Galveston approved this study.
Study Population and Sample Size Needed
Our study is based on the analytical SEER–Medicare files that excluded women who did not have full coverage of both Medicare Part A and Part B and those who were members of health maintenance organizations, because claims from these organizations may not be complete. The study population consisted of 6190 women aged ≥65 who were diagnosed with operable breast cancer (stage II or IIIA) and also with positive lymph nodes during the period from 1992 to 1996, identified from the 11 SEER areas. Of the total study population of 6190 women, 715 (11.6%) who had other primary cancers were excluded. Also excluded were 11 women who died within 3 months of breast cancer diagnosis. Thus, 5464 patients aged ≥65 remained for the final analyses. Assuming a difference of 8% in survival, the sample size needed to achieve 80% power at p < .05 is 1139 (aged 65–60 years) and 2904 (aged 70 or older) (20).
Study Variables
Chemotherapy
The details of the methods of identification of chemotherapy use through Medicare claims have been previously described (21,22). In brief, patients with breast cancer were defined as having received chemotherapy if any of the Medicare procedure codes indicated so within 6 months of the diagnosis (21–24).
Comorbidity index
Comorbidity was ascertained from Medicare claims data through diagnoses or procedures made 1 year prior to and 1 month after the diagnosis of breast cancer, but not including the breast cancer diagnosis. We included claims for comorbid diseases made within 1 month of diagnosis because any comorbid diseases present at that time would be likely to affect the choice of chemotherapy. We adopted the comorbidity index created by Charlson and colleagues (25), and later validated by Romano and colleagues (26), using the International Classification of Diseases, 9th revision (ICD-9-CM) diagnosis and procedure codes. Women diagnosed with breast cancer at age 65 may have a shorter duration for identifying comorbid diseases in Medicare, but these women had similar comorbidity scores to those aged 66–69 years, and therefore were included in the analysis.
Mortality
For patients who died of any causes after diagnosis of breast cancer (all-cause mortality), information on months of survival from the date of diagnosis was provided in the SEER database. The last date of the follow-up for this cohort was December 31, 1999.
Other characteristics
Patient and tumor characteristics such as age, race/ethnicity, marital status, tumor stage, tumor size, tumor grade, hormone receptor status, surgery, radiotherapy, and geographic areas (11 SEER areas) were available from the SEER data. Information on household income at the level of the census tract in the SEER–Medicare database was obtained from the National Census Bureau.
Analyses
The Cox proportional hazard ratio (HR) of mortality was analyzed using the PHREG procedure after patients who were lost to follow-up or who were still alive at the last date of follow-up were censored, adjusting for patient and tumor characteristics that are shown and categorized in Table 1 (27,28).
Table 1.
Comparison of Characteristics of Older Women With Node-Positive Operable Breast Cancer Who Received Adjuvant Chemotherapy Versus Those Who Did Not, Before and After Adjusting for Propensity Score
| Comparison in Column Percentage Between Patients Receiving Chemotherapy and Those Not Receiving Chemotherapy | ||||||
|---|---|---|---|---|---|---|
| Patient and Tumor Characteristics | No. of Patients With Breast Cancer | Percentage of Women Receiving Adjuvant Chemotherapy | Patients Not Receiving Chemotherapy
(N = 3774) |
Patients Receiving Chemotherapy
(N = 1690) |
p Value Before Adjustment* | p Value After Adjusting for Propensity Score† |
| Age, y | ||||||
| 65–69 | 1440 | 53.5 | 17.7 | 45.6 | <.001 | .018 |
| 70–74 | 1590 | 36.1 | 26.9 | 34.0 | ||
| 75–79 | 1275 | 20.1 | 27.0 | 15.2 | ||
| 80–84 | 753 | 10.4 | 17.9 | 4.6 | ||
| ≥85 | 406 | 2.7 | 10.5 | 0.7 | ||
| Race | ||||||
| White | 4916 | 30.7 | 90.3 | 89.2 | .209 | .995 |
| Black | 363 | 31.4 | 6.6 | 6.8 | ||
| Other | 185 | 36.8 | 3.1 | 4.0 | ||
| Marital status | ||||||
| Married | 2365 | 38.1 | 38.8 | 53.4 | <.001 | .875 |
| Unmarried | 2998 | 24.9 | 59.7 | 44.1 | ||
| Unknown | 101 | 41.6 | 1.6 | 2.5 | ||
| Tumor stage | ||||||
| Stage II | 4694 | 30.0 | 87.1 | 83.2 | <.001 | .058 |
| Stage IIIA | 770 | 36.9 | 12.9 | 16.8 | ||
| Tumor size, cm | ||||||
| <1.0 | 311 | 31.2 | 5.7 | 5.7 | <.001 | .729 |
| 1.0–<2.0 | 1570 | 27.1 | 30.3 | 25.2 | ||
| 2.0–<3.0 | 1527 | 31.2 | 27.9 | 28.2 | ||
| 3.0–<4.0 | 820 | 32.2 | 14.7 | 15.6 | ||
| 4.0+ | 1005 | 33.9 | 17.6 | 20.2 | ||
| Unknown size | 231 | 37.7 | 3.8 | 5.2 | ||
| No. of positive nodes | ||||||
| 1 | 1939 | 22.5 | 41.3 | 22.5 | <.001 | .792 |
| 2–4 | 1908 | 32.8 | 35.9 | 32.8 | ||
| 5–9 | 879 | 22.5 | 13.2 | 22.5 | ||
| 10–43 | 668 | 20.4 | 8.6 | 20.4 | ||
| Unknown No., but positive | 70 | 1.8 | 1.1 | 1.8 | ||
| Tumor grade | ||||||
| Well differentiated | 400 | 18.3 | 8.7 | 4.3 | <.001 | .782 |
| Moderately differentiated | 1836 | 25.7 | 36.1 | 27.9 | ||
| Poorly differentiated | 1909 | 38.7 | 31.0 | 43.7 | ||
| Unknown | 1319 | 30.8 | 24.2 | 24.0 | ||
| Hormone receptor status | ||||||
| Positive | 3786 | 25.9 | 74.3 | 58.1 | <.001 | .007 |
| Negative | 779 | 58.3 | 8.6 | 26.9 | ||
| Unknown | 899 | 28.4 | 17.1 | 15.1 | ||
| Surgery and radiation | ||||||
| BCS alone | 290 | 33.8 | 5.1 | 5.8 | <.001 | .354 |
| BCS with radiation | 1001 | 30.6 | 18.4 | 18.1 | ||
| Mastectomy alone | 3579 | 28.8 | 67.5 | 61.0 | ||
| Mastectomy with radiation | 594 | 42.9 | 9.0 | 15.1 | ||
| Comorbidity scores | ||||||
| 0 | 3725 | 32.9 | 66.2 | 72.5 | <.001 | .489 |
| 1 | 658 | 27.5 | 12.6 | 10.7 | ||
| 2 | 222 | 18.9 | 4.8 | 2.5 | ||
| ≥3 | 859 | 28.2 | 16.4 | 14.3 | ||
| Household income | ||||||
| 1st quartile (≤ $27669) | 1080 | 28.1 | 20.6 | 17.9 | <.001 | .796 |
| 2nd quartile ($27,670–$34,464) | 1107 | 29.5 | 20.7 | 19.3 | ||
| 3rd quartile ($34,465–$43,974) | 1007 | 33.8 | 17.7 | 20.1 | ||
| 4th quartile (>$43,974) | 1002 | 35.1 | 17.2 | 20.8 | ||
| Unknown | 1268 | 29.1 | 23.8 | 21.8 | ||
| SEER registry areas | ||||||
| San Francisco, CA | 465 | 26.9 | 9.0 | 7.4 | <.001 | .906 |
| Connecticut | 790 | 36.8 | 13.2 | 17.2 | ||
| Detroit, MI | 896 | 33.9 | 15.7 | 18.0 | ||
| Hawaii | 106 | 39.6 | 1.7 | 2.5 | ||
| Iowa | 751 | 27.2 | 14.5 | 12.1 | ||
| New Mexico | 212 | 22.6 | 4.4 | 2.8 | ||
| Seattle, WA | 590 | 29.7 | 11.0 | 10.4 | ||
| Utah | 270 | 30.4 | 5.0 | 4.9 | ||
| Atlanta, GA | 307 | 32.3 | 5.5 | 5.9 | ||
| San Jose–Monterey, CA | 259 | 32.1 | 4.7 | 4.9 | ||
| Los Angeles, CA | 828 | 29.0 | 15.4 | 14.0 | ||
| Year of diagnosis | ||||||
| 1992 | 1203 | 27.9 | 23.0 | 19.9 | <.01 | .775 |
| 1993 | 1128 | 28.3 | 21.4 | 18.9 | ||
| 1994 | 1114 | 31.7 | 20.2 | 20.9 | ||
| 1995 | 1083 | 32.7 | 19.3 | 21.0 | ||
| 1996 | 936 | 35.0 | 16.1 | 19.4 | ||
| Total | 5464 | 30.9 | 100.0 | 100.0 | ||
p value was generated from the Mantel—Haenszel chi-square statistic.
p value was generated from the Cochran Mantel–Haenszel chi-square statistic, adjusting for propensity score (categorized in quintiles).
BCS = breast-conserving surgery; SEER = The National Cancer Institute's Surveillance, Epidemiology, and End Results cancer registry.
A propensity score analysis was also conducted to minimize selection bias in this observational study (29,30). In observational studies, investigators have no control over the treatment assignment. Patients in the treated and untreated groups may have large differences in their observed covariates, and these differences can lead to biased estimates of treatment effects. Propensity score can be used to balance the covariates between the treated and untreated groups, and therefore reduce this bias (30). Propensity score was defined as the conditional probability of being treated, given the covariates. The probability that a patient would receive chemotherapy was generated from the logistic regression model that incorporated potential confounding factors listed in Table 1. Propensity score was then divided into five strata (quintiles) for all patients and stratified for age groups. It is estimated that five strata of propensity scores are able to remove over 90% of the unbalanced bias from confounding factors between the treated and untreated groups (30). In this study, the Mantel–Haenszel chi-square test was used to initially test for any significant differences in the distribution of confounding factors between the treated and untreated groups. The balances of confounding factors were tested again using the Cochran Mantel–Haenszel chi square, which adjusted for propensity scores (in quintiles). The covariates that were still significantly different between the treated and untreated groups were adjusted again, together with propensity score in the Cox proportional hazard models. In addition, survival analysis for the HR of mortality was also performed to show the effect of other factors on mortality.
Sensitivity analyses were performed to assess the potential effects of unmeasured confounders on the observed associations between chemotherapy and mortality. The unmeasured confounders in our study could be patient level covariates, such as patient preference, and physician/clinic level covariates, such as oncologist clustering in preference to using chemotherapy. Sensitivity analyses were conducted using the method of Lin and colleagues (31). The prevalence of the unmeasured confounder was set to different levels within both the treated and untreated groups. Over the different prevalence levels, HR of the unmeasured confounder and mortality ranged from 1.25 to 2.50. The range for the prevalence estimates and for the HRs was obtained by inspection of the prevalence and hazard associated with the measured confounders in this study. The Cox regression parameter for chemotherapy was adjusted to determine the effect of the unknown confounder on HR in the treated group compared to the untreated group. All computer programming and analyses were completed using the SAS system (28).
Results
Table 1 shows the percentage of women with node-positive operable breast cancer receiving adjuvant chemotherapy by various patient and tumor characteristics, and the comparison between patients who received chemotherapy and those who did not receive chemotherapy in terms of distribution of these characteristics, before and after adjusting for propensity score. Overall, 31% of women with node-positive operable breast cancer received adjuvant chemotherapy. The use of adjuvant chemotherapy decreased substantially with age, from 54% in women aged 65–69 to 3% in women aged ≥85. The distribution of cases between the treated and untreated groups varied significantly by potential confounding factors (except ethnicity) listed in the table. After adjusting for propensity to receive chemotherapy, the treated and untreated groups were not statistically significantly different in covariates except for age and hormone receptor status.
Table 2 presents HR for mortality associated with receipt of chemotherapy for all older women with breast cancer and for the five age groups. HR was statistically significantly reduced in women aged 65–69 years who received adjuvant chemotherapy compared to those who did not receive chemotherapy (0.70, 95% confidence interval [CI], 0.57–0.88), after adjusting for other factors that might affect survival. HR did not significantly differ between the treated and untreated groups in women aged ≥70.
Table 2.
Effect of Chemotherapy on Mortality in Women With Node-Positive Operable Breast Cancer Diagnosed in 1992–1996 in SEER Areas
| Age Group | Total No. of Patients (% of Patients Receiving Chemotherapy) | Risk (Cox Proportional HR and 95% CI)* | |
|---|---|---|---|
| HR | 95% CI | ||
| 65–69 y | 1440 (53.5) | 0.70 | 0.57–0.88 |
| 70–74 y | 1590 (36.1) | 0.87 | 0.70–1.07 |
| 75–79 y | 1275 (20.1) | 1.06 | 0.84–1.34 |
| 80–84 y | 753 (10.4) | 0.90 | 0.63–1.29 |
| ≥85 y | 406 (2.7) | 1.21 | 0.53–2.74 |
| All women aged ≥65 | 5464 (30.9) | 0.90 | 0.80–1.01 |
Cases who received chemotherapy vs those who did not. Cox proportional HR was adjusted for age (continuous), ethnicity (white, black, other), marital status (married, unmarried, unknown), tumor stage (II, IIIA), tumor size (<1, 1–<2, 2–<3, 3–<4, ≥4 cm, or unknown), hormone receptor status (positive, negative, or unknown), tumor grade (well, moderately or poorly differentiated, or undetermined), number of positive lymph nodes (1, 2–4, 5–9, 10–43, and positive nodes but unknown number), comorbidity index scores (0, 1, 2, or 3+), 11 SEER areas, other treatment (breast-conserving surgery alone, breast-conserving surgery plus radiotherapy, mastectomy alone, or mastectomy with radiation), and census tract household income in quartiles (four categories plus unknown one), and year of diagnosis (continuous).
SEER = The National Cancer Institute's Surveillance, Epidemiology, and End Results cancer registry; HR = hazard ratio; CI = confidence interval.
Table 3 presents HR stratified for two separate age groups (65–69 and ≥70) to show the effects of chemotherapy and other potential confounders on mortality. As shown above, among the 1440 women aged 65–69, those who received adjuvant chemotherapy had a significantly reduced mortality when compared to those who did not receive chemotherapy. Mortality was also significantly associated with tumors of larger size, more positive lymph nodes, higher tumor grade, negative hormone receptor, higher comorbidity scores, and being unmarried. Among 4024 women aged ≥70, the effect of chemotherapy on mortality was not statistically significant.
Table 3.
Effect of Chemotherapy and Other Factors on Mortality in Women With Node-Positive Operable Breast Cancer Diagnosed in 1992–1996 in SEER Areas, by Age Groups (65–69 Years and 70+ Years)
| Risk (Cox Proportional HR and 95% CI)* | ||
|---|---|---|
| Patient and Tumor Characteristics | Women Aged 65–69 Years (N = 1440) | Women Aged 70 Years or Older (N = 4024) |
| Chemotherapy | ||
| No | 1.00 (reference) | 1.00 (reference) |
| Yes | 0.70 (0.57–0.88) | 0.96 (0.83–1.09) |
| Age (continuous) | 0.96 (0.90–1.04) | 1.06 (1.05–1.07) |
| Race | ||
| White | 1.00 (reference) | 1.00 (reference) |
| Black | 1.19 (0.85–1.67) | 1.12 (0.90–1.39) |
| Other | 1.56 (0.87–2.77) | 0.85 (0.59–1.23) |
| Marital status | ||
| Married | 1.00 (reference) | 1.00 (reference) |
| Unmarried | 1.36 (1.10–1.67) | 1.33 (1.19–1.49) |
| Unknown | 1.29 (0.69–2.42) | 1.44 (1.01–2.04) |
| Tumor stage | ||
| II | 1.00 (reference) | 1.00 (reference) |
| IIIA | 1.00 (0.73–1.37) | 1.20 (1.04–1.39) |
| Tumor size, cm | ||
| <1.0 | 1.00 (reference) | 1.00 (reference) |
| 1.0–<2.0 | 1.00 (0.61–1.63) | 1.20 (0.89–1.62) |
| 2.0–<3.0 | 1.26 (0.78–2.04) | 1.58 (1.18–2.12) |
| 3.0–<4.0 | 1.32 (0.79–2.21) | 1.99 (1.47–2.69) |
| 4.0+ | 2.00 (1.20–3.32) | 1.91 (1.41–2.59) |
| Unknown size | 1.35 (0.72–2.53) | 1.34 (0.92–1.94) |
| Number of positive nodes | ||
| 1 | 1.00 (reference) | 1.00 (reference) |
| 2–4 | 1.30 (0.99–1.71) | 1.11 (0.98–1.25) |
| 5–9 | 2.07 (1.52–2.80) | 1.38 (1.19–1.61) |
| 10–43 | 2.56 (1.86–3.52) | 1.94 (1.65–2.28) |
| Unknown no. but positive | 2.12 (0.99–4.55) | 1.72 (1.19–2.48) |
| Tumor grade | ||
| Well differentiated | 1.00 (reference) | 1.00 (reference) |
| Moderately differentiated | 1.38 (0.82–2.31) | 0.99 (0.78–1.25) |
| Poorly differentiated | 1.77 (1.06–2.97) | 1.30 (1.03–1.65) |
| Unknown | 1.46 (0.86–2.47) | 1.14 (0.89–1.44) |
| Hormone receptor status | ||
| Positive | 1.00 (reference) | 1.00 (reference) |
| Negative | 2.13 (1.65–2.74) | 1.76 (1.53–2.02) |
| Unknown | 1.02 (0.76–1.36) | 1.22 (1.07–1.39) |
| Surgery and radiation | ||
| BCS alone | 1.00 (reference) | 1.00 (reference) |
| BCS with radiation | 0.61 (0.38–1.00) | 0.81 (0.62–1.05) |
| Mastectomy alone | 0.79 (0.51–1.21) | 0.94 (0.75–1.18) |
| Mastectomy with radiation | 0.82 (0.50–1.34) | 0.91 (0.70–1.19) |
| Comorbidity scores | ||
| 0 | 1.00 (reference) | 1.00 (reference) |
| 1 | 1.13 (0.81–1.57) | 1.20 (1.03–1.38) |
| 2 | 1.78 (1.06–2.96) | 1.28 (1.03–1.58) |
| ≥3 | 1.42 (1.08–1.86) | 1.16 (1.02–1.31) |
| Household income | ||
| 1st quartile (≤$27,669) | 1.00 (reference) | 1.00 (reference) |
| 2nd quartile ($27,670–$34,464) | 1.20 (0.87–1.65) | 0.98 (0.83–1.14) |
| 3rd quartile ($34,465–$43,974) | 1.06 (0.73–1.53) | 0.93 (0.78–1.11) |
| 4th quartile (>$43,974) | 1.20 (0.81–1.76) | 1.01 (0.83–1.23) |
| Unknown | 1.52 (0.86–2.70) | 1.22 (0.95–1.58) |
| Total | ||
Mortality among women with node-positive operable breast cancer, by age. Cox proportional HR was adjusted for those variables listed, plus 11 SEER areas (categorized) and year of diagnosis (continuous).
SEER = The National Cancer Institute's Surveillance, Epidemiology, and End Results cancer registry; HR = hazard ratio; CI = confidence interval; BCS = breast-conserving surgery.
We performed an analysis regarding the effect of chemotherapy on mortality by adjusting for propensity score. Those covariates (i.e., age and hormone receptor status) that remained statistically unbalanced between the treated and untreated groups as shown in Table 1 were also adjusted together with propensity score in the analysis. Mortality was still statistically significantly reduced in women aged 65–69 who received chemotherapy compared to those who did not (HR = 0.76, 95% CI, 0.62–0.94). Chemotherapy did not have a statistically significant effect on mortality in women with breast cancer aged ≥70 (HR = 0.99, 95% CI, 0.87–1.14).
We also conducted sensitivity analyses to estimate the potential effect of unknown confounders on the study results (Table 4). The estimated prevalence and HR of an unknown factor was based on the prevalence and HR of measured variables shown in Table 1 and Table 3. For example, 53% of women receiving chemotherapy were married compared to 39% of those not receiving chemotherapy, a 14% difference between the treated and untreated groups (Table 1). HR of mortality was 1.36 (95% CI, 1.10–1.67) in unmarried women aged 65–69 compared to married women (Table 3). The highest HR associated with a measured confounder was about 2.50 for women with ≥10 positive nodes (Table 3). Based on these findings from measured variables, we set the prevalence of the unmeasured confounder up to 30% difference level between the treated and untreated groups. Over the different prevalence levels, the HR of the unmeasured confounder for mortality ranged from 1.25 to 2.50. These analyses show that an unmeasured confounder could be responsible for the results obtained if it has difference of approximately 30% in prevalence between the treated and nontreated groups and if it is associated with an HR of 2.00 or greater for patients aged 65–69 and those aged 70 or older.
Table 4.
Subset of Sensitivity Analyses of the Hazard Ratio of Mortality for Women Receiving Chemotherapy Compared to Those Receiving No Chemotherapy, Controlling for an Unmeasured Binary Confounder
| Prevalence (%) of Unknown Confounder in the Chemotherapy Group | Prevalence (%) of Unknown Confounder in the Untreated Group | Unknown Confounder Hazard Ratio | Hazard Ratio of Mortality, Adjusted for Unmeasured Confounder (95% Confidence Interval) | |
|---|---|---|---|---|
| Age 65–69 Years | Age 70 Years or Older | |||
| 0 | 0 | 1.00 | 0.70 (0.56–0.88) | 0.96 (0.83–1.09) |
| 50 | 70 | 1.25 | 0.73 (0.59–0.92) | 1.00 (0.87–1.14) |
| 40 | 70 | 1.25 | 0.75 (0.60–0.94) | 1.02 (0.89–1.17) |
| 50 | 70 | 1.50 | 0.76 (0.61–0.95) | 1.03 (0.90–1.18) |
| 40 | 70 | 1.50 | 0.79 (0.64–0.99) | 1.07 (0.94–1.23) |
| 50 | 70 | 2.00 | 0.80 (0.64–0.99) | 1.08 (0.95–1.24) |
| 40 | 70 | 2.00 | 0.85 (0.69–1.06) | 1.16 (1.01–1.33) |
| 50 | 70 | 2.50 | 0.82 (0.66–1.03) | 1.12 (0.98–1.28) |
| 40 | 70 | 2.50 | 0.90 (0.72–1.12) | 1.22 (1.07–1.40) |
Discussion
This study examined the effectiveness of adjuvant chemotherapy in 5464 women aged 65 or older with node-positive operable breast cancer. Mortality was significantly reduced in women aged 65–69 years who received adjuvant chemotherapy compared to those who did not, after adjusting for measured confounders and propensity score. However, chemotherapy failed to show significant effect on mortality in women aged ≥70.
There are several strengths to this study. First, the data were obtained from the National Cancer Institute's SEER program, which accounts for 14% of the U.S. population and ascertains nearly all cancer patients in the designated SEER areas (15). The advantage of this population coverage is that the data include all women residing in community-based settings, which avoids selection bias such as volunteer bias in randomized trials. Although 46% of incident breast cancer cases occur in women aged ≥65 and 32% of new cases are women aged ≥70 (15), only a small fraction of older women are enrolled in the randomized trials (1–3). For example, in the 47 trials on chemotherapy, 18,718 women were recruited, of whom only 609 (3.3%) were aged ≥70 (3).
Furthermore, because randomized clinical trials often have tight enrollment criteria and are conducted in motivated volunteers, one might expect that the effectiveness of chemotherapy in the community at large might be different from the efficacy of chemotherapy found in the controlled trials. In fact, to the best of our knowledge there is no study that has assessed the effectiveness of chemotherapy for breast cancer in routine practice.
Age has been demonstrated to be a major factor in determining cancer therapies, particularly for chemotherapy, because there are concerns about comorbidity and drug toxicity (21–23,32–40). However, for women with breast cancer, there might be an additional reason for the declining use of chemotherapy with age, i.e., the decreasing efficacy of chemotherapy with age for breast cancer could have affected the knowledge and attitudes of oncologists and patients on the benefits and risks of chemotherapy (34). The decrease in both efficacy and effectiveness of chemotherapy with advanced age in women with breast cancer is not seen in men and women with colon cancer (34,35). In those persons with node-positive resectable colon cancer, chemotherapy seems equally efficacious and effective in patients aged ≥70 compared to those aged <70 (35).
It is important to note several weaknesses of this study. First, there is concern regarding whether the information on chemotherapy from Medicare claims is accurate and complete. However, a recent study comparing Medicare claims with medical chart reviews on the use of chemotherapy showed excellent agreement rates (ranging from 97.0% to 98.3%) (41). An additional limitation is that the study findings may only be applied to women aged ≥65 who are not members of health maintenance organizations and have both Medicare Part A and Part B coverage. Furthermore, there was no information on the dose intensity of chemotherapy. Because of concern about lower tolerance of this therapy in older women, a lower dose of chemotherapy might have been given, which could have been less effective, thus confounding the results. In addition, we do not have information on adjuvant hormone therapy use, which may impact survival in older women with breast cancer. In a hospital-based study in several states across the United States, 59% of women with early-stage breast cancer aged 65 or older who received adjuvant chemotherapy also received adjuvant hormone therapy compared to 79% of women who did not receive adjuvant chemotherapy, a 20% difference (42). This difference of receiving hormone therapy among women who received chemotherapy and those who did not in the community setting was much smaller (33). Adjuvant tamoxifen therapy can achieve up to a 26% reduction in mortality (43). Our sensitivity analyses showed that the effect of adjuvant chemotherapy on mortality would not be much affected by confounding with above prevalence and magnitude of effect.
Women with breast cancer who received chemotherapy may have a different spectrum of prognostic indicators than those who did not. As shown in Table 1, women receiving chemotherapy had a higher percentage of large-sized tumors, more positive lymph nodes, and poorly differentiated tumors. These worse prognostic factors would have led to bias towards poorer survival. In contrast, women receiving chemotherapy had fewer coexisting diseases, were more likely to be married, and had a relatively higher socioeconomic status than those not receiving chemotherapy (Table 1); these factors would have led to better survival. Patients aged 65–69 had a higher percentage of anthracycline agent use (43%) than did those patients aged 70 or older (32%), but mortality was not affected because the HRs for anthracycline versus nonanthracycline agents were identical (HR = 0.99, 95% CI, 0.83–1.18) in our study. Although the differential distribution of these prognostic factors was minimized through propensity scores, there still might be some factors not linked to the measured confounders that affected the results. The sensitivity analyses suggested that such an unmeasured confounder would have to have a substantial effect on survival and a large difference in distribution between the chemotherapy and nonchemotherapy groups to affect the results. Also, it is difficult to envision how such a confounder or set of confounders could have produced the pattern of effectiveness in 65- to 69-year-olds with no effectiveness in those aged 70 or older. Because there are more diverse characteristics in older patients, adopting sensitivity analysis to estimate the potential effects of unmeasured or unknown factors are useful and should be encouraged in the studies of older individuals.
Conclusion
Our study showed that adjuvant chemotherapy is associated with improved survival in women with node-positive operable breast cancer aged 65–69 living in the community, but there is no proof that adjuvant chemotherapy reduces the risk of mortality in women aged 70 or older. These findings were consistent with those found in pooled analyses of randomized controlled trials (1–3). Further study may be needed to determine whether these findings regarding the effectiveness of chemotherapy in routine practices remain the same in different populations and in various communities. Particularly because there is no proof that a 70-year-old woman is biologically different from a 68- or 69-year-old, further observational studies and randomized trials with sufficient sample size would be needed to generate the definitive answer and to identify those aged 70 or older who may benefit from adjuvant chemotherapy, if any. The answers to the above research questions would have important clinical implications for oncologists and physicians in treating older women with breast cancer, for the National Institutes of Health consensus conferences in developing clinical guidelines for chemotherapy, and for the benefit/risk assessment of this therapy by physicians and patients themselves.
Acknowledgments
This study was supported by grants from the National Cancer Institute (R01-CA90626 to Dr. Du), and the Department of Defense (DAMD17-99-1-9397 to Dr. Du).
We thank Jonathan Mahnkan, MS, for help with sensitivity analysis.
This study used the Linked SEER-Medicare Database.
The interpretation and reporting of these data are the sole responsibilities of the authors.
The authors acknowledge the efforts of the Applied Research Branch, Division of Cancer Prevention and Population Science, National Cancer Institute; the Office of Information Services, and the Office of Strategic Planning, Center for Medicare and Medicaid Services; Information Management Services, Inc.; and the SEER Program tumor registries in the creation of the SEER–Medicare database.
References
- 1.Early Breast Cancer Trialists' Collaborative Group. Effects of adjuvant tamoxifen and of cytotoxic therapy on mortality in early breast cancer: an overview of 61 randomised trials among 28,896 women. N Engl J Med. 1988;319:1681–1692. doi: 10.1056/NEJM198812293192601. [DOI] [PubMed] [Google Scholar]
- 2.Early Breast Cancer Trialists' Collaborative Group. Systemic treatment of early breast cancer by hormonal, cytotoxic, or immune therapy: 133 randomised trials involving 31,000 recurrences and 24,000 deaths among 75,000 women. Lancet. 1992;339:1–15. 71–85. [PubMed] [Google Scholar]
- 3.Early Breast Cancer Trialists' Collaborative Group. Polychemotherapy for early breast cancer: an overview of the randomised trials. Lancet. 1998;352:930–942. [PubMed] [Google Scholar]
- 4.Fisher B, Brown AM, Dimitrov NV, et al. Two months of doxorubicin-cyclophosphamide with and without interval reinduction therapy compared with 6 months of cyclophosphamide, methotrexate, and fluorouracil in positive-node breast cancer patients with tamoxifen-nonresponsive tumors: results from the National Surgical Adjuvant Breast and Bowel Project B-15. J Clin Oncol. 1990;8:1483–1496. doi: 10.1200/JCO.1990.8.9.1483. [DOI] [PubMed] [Google Scholar]
- 5.Cole BF, Gelber RD, Gelber S, Coates AS, Goldhirsch A. Polychemotherapy for early breast cancer: an overview of the randomised clinical trials with quality-adjusted survival analysis. Lancet. 2001;358:277–286. doi: 10.1016/S0140-6736(01)05483-6. [DOI] [PubMed] [Google Scholar]
- 6.NIH Consensus Conference. Adjuvant chemotherapy for breast cancer. JAMA. 1985;254:3461–3463. [PubMed] [Google Scholar]
- 7.NIH Consensus Conference. Treatment of early-stage breast cancer. JAMA. 1991;265:391–395. [PubMed] [Google Scholar]
- 8.The Steering Committee on Clinical Practice Guidelines for the Care and Treatment of Breast Cancer. Adjuvant systemic therapy for women with node-negative breast cancer. Can Med Assoc J. 1998;158(Suppl 3):S43–S51. [PubMed] [Google Scholar]
- 9.The Steering Committee on Clinical Practice Guidelines for the Care and Treatment of Breast Cancer. Adjuvant systemic therapy for women with node-positive breast cancer. Can Med Assoc J. 1998;158(Suppl 3):S52–S64. [PubMed] [Google Scholar]
- 10.National Institutes of Health Consensus Development Panel. National Institutes of Health Consensus Development Conference Statement: Adjuvant Therapy for Breast Cancer, November 1–3, 2000. J Natl Cancer Inst. 2001;93:979–989. doi: 10.1093/jnci/93.13.979. [DOI] [PubMed] [Google Scholar]
- 11.Black N. Why we need observational studies to evaluate the effectiveness of health care. BMJ. 1996;312:1215–1218. doi: 10.1136/bmj.312.7040.1215. [DOI] [PMC free article] [PubMed] [Google Scholar]
- 12.Benson K, Hartz AJ. A comparison of observational studies and randomized, controlled trials. N Engl J Med. 2000;342:1878–1886. doi: 10.1056/NEJM200006223422506. [DOI] [PubMed] [Google Scholar]
- 13.Concato J, Shah N, Horwitz RI. Randomized, controlled trials, observational studies, and the hierarchy of research designs. N Engl J Med. 2000;342:1887–1892. doi: 10.1056/NEJM200006223422507. [DOI] [PMC free article] [PubMed] [Google Scholar]
- 14.Pocock SJ, Elbourne DR. Randomized trials or observational tribulations? N Engl J Med. 2000;342:1907–1909. doi: 10.1056/NEJM200006223422511. [DOI] [PubMed] [Google Scholar]
- 15.Ries LAG, Kosary CL, Hankey BF, et al. SEER Cancer Statistics Review 1973–1996. NIH Publication No 99-2789. Bethesda, MD: National Cancer Institute; 1999. [Google Scholar]
- 16.National Cancer Institute. The SEER Program Code Manual. Revised. Bethesda, MD: National Cancer Institute; 1994. NIH Publication No 94-1999. [Google Scholar]
- 17.Potosky AL, Riley GF, Lubitz JD, et al. Potential for cancer related health services research using a linked Medicare-tumor registry database. Med Care. 1993;31:732–748. [PubMed] [Google Scholar]
- 18.Health Care Financing Administration. Data User Reference Guide (DURG) Baltimore, MD: Health Care Financing Administration; Jan, 2000. [Google Scholar]
- 19.Warren JL, Klabunde CN, Schrag D, Bach PB, Riley GF. Overview of the SEER-Medicare data: content, research applications, and generalizability to the United States elderly population. Med Care. 2002;40(Suppl):IV3–18. doi: 10.1097/01.MLR.0000020942.47004.03. [DOI] [PubMed] [Google Scholar]
- 20.Jerry Hintze. [May 1, 2004];NCSS – Statistical and Power Analysis Software-PASS 2003. Available at: www.ncss.com.
- 21.Du XL, Goodwin JS. Patterns of use of chemotherapy for breast cancer in older women: findings from Medicare claims data. J Clin Oncol. 2001;19:1455–1461. doi: 10.1200/JCO.2001.19.5.1455. [DOI] [PubMed] [Google Scholar]
- 22.Du XL, Goodwin JS. Increase of chemotherapy use in older women with breast carcinoma from 1991 to 1996. Cancer. 2001;92:730–737. doi: 10.1002/1097-0142(20010815)92:4<730::aid-cncr1376>3.0.co;2-p. [DOI] [PMC free article] [PubMed] [Google Scholar]
- 23.Du XL, Osborne C, Goodwin JS. Population-based assessment of hospitalizations for toxicity from chemotherapy in older women with breast cancer. J Clin Oncol. 2002;20:4636–4642. doi: 10.1200/JCO.2002.05.088. [DOI] [PMC free article] [PubMed] [Google Scholar]
- 24.Health Care Financing Administration. HCFA Common Procedure Coding System (HCPCS): National Level II Medicare Codes. Los Angeles, CA: Practice Management Information Corporation; 1994. [Google Scholar]
- 25.Charlson ME, Pompei P, Ales KL, et al. A new method of classifying prognostic comorbidity in longitudinal studies: development and validation. J Chronic Dis. 1987;40:373–383. doi: 10.1016/0021-9681(87)90171-8. [DOI] [PubMed] [Google Scholar]
- 26.Romano PS, Roos LL, Jollis JG. Adapting a clinical comorbidity index for use with ICD-9-CM administrative data: differing perspectives. J Clin Epidemiol. 1993;46:1075–1079. doi: 10.1016/0895-4356(93)90103-8. [DOI] [PubMed] [Google Scholar]
- 27.Cox DR. Regression models and life tables. J R Stat Soc. 1972;34:187–202. [Google Scholar]
- 28.SAS Institute Inc. SAS/STAT Software: Changes and Enhancements. Cary, NC: SAS Institute, Inc.; 1996. [Google Scholar]
- 29.Rosenbaum PR, Rubin DB. The central role of the propensity score in observational studies for causal effects. Biometrika. 1983;70:41–55. [Google Scholar]
- 30.D'Agostino RB., Jr Propensity score methods for bias reduction in the comparison of a treatment to a non-randomized control group. Stat Med. 1998;17:2265–2281. doi: 10.1002/(sici)1097-0258(19981015)17:19<2265::aid-sim918>3.0.co;2-b. [DOI] [PubMed] [Google Scholar]
- 31.Lin DY, Psaty BM, Kronmal RA. Assessing the sensitivity of regression results to unmeasured confounders in observational studies. Biometrics. 1998;54:948–963. [PubMed] [Google Scholar]
- 32.Yancik R, Wesley MN, Ries LA, Havlik RJ, Edwards BK, Yates JW. Effect of age and comorbidity in postmenopausal breast cancer patients aged 55 years and older. JAMA. 2001;285:885–892. doi: 10.1001/jama.285.7.885. [DOI] [PubMed] [Google Scholar]
- 33.Du XL, Key CR, Osborne C, et al. Discrepancy between consensus recommendations and actual community use of adjuvant chemotherapy in women with breast cancer. Ann Intern Med. 2003;138:90–97. doi: 10.7326/0003-4819-138-2-200301210-00009. [DOI] [PMC free article] [PubMed] [Google Scholar]
- 34.Sargent DJ, Goldberg RM, Jacobson SD, et al. A pooled analysis of adjuvant chemotherapy for resected colon cancer in elderly patients. N Engl J Med. 2001;345:1091–1097. doi: 10.1056/NEJMoa010957. [DOI] [PubMed] [Google Scholar]
- 35.Muss HB. Older age—not a barrier to cancer treatment. N Engl J Med. 2001;345:1127–1128. doi: 10.1056/nejm200110113451512. [DOI] [PubMed] [Google Scholar]
- 36.Begg CB, Cohen JL, Ellerton J. Are the elderly predisposed to toxicity from cancer chemotherapy? An investigation using data from the Eastern Cooperative Oncology Group. Cancer Clin Trials. 1980;3:369–374. [PubMed] [Google Scholar]
- 37.Shapiro CL, Recht A. Drug therapy: side effects of adjuvant treatment of breast cancer. N Engl J Med. 2001;344:1997–2008. doi: 10.1056/NEJM200106283442607. [DOI] [PubMed] [Google Scholar]
- 38.Harlan LC, Abrams J, Warren JL, Clegg L, Stevens J, Ballard-Barbash R. Adjuvant therapy for breast cancer: practice patterns of community physicians. J Clin Oncol. 2002;20:1809–1817. doi: 10.1200/JCO.2002.07.052. [DOI] [PubMed] [Google Scholar]
- 39.Mariotto A, Feuer EJ, Harlan LC, Wun LM, Johnson KA, Abrams J. Trends in use of adjuvant multi-agent chemotherapy and tamoxifen for breast cancer in the United States: 1975–1999. J Natl Cancer Inst. 2002;94:1626–1634. doi: 10.1093/jnci/94.21.1626. [DOI] [PubMed] [Google Scholar]
- 40.Du XL. Re: Trends in use of adjuvant multi-agent chemotherapy and tamoxifen for breast cancer in the United States: 1975–1999. J Natl Cancer Inst. 2003;95:683–685. doi: 10.1093/jnci/95.9.683. [DOI] [PubMed] [Google Scholar]
- 41.Warren JL, Harlan LC, Fahey A, et al. Utility of the SEER-Medicare data to identify chemotherapy use. Med Care. 2002;40(Suppl):IV55–61. doi: 10.1097/01.MLR.0000020944.17670.D7. [DOI] [PubMed] [Google Scholar]
- 42.Silliman RA, Guadagnoli E, Rakowski W, et al. Adjuvant tamoxifen prescription in women 65 years and older with primary breast cancer. J Clin Oncol. 2002;20:2680–2688. doi: 10.1200/JCO.2002.08.137. [DOI] [PubMed] [Google Scholar]
- 43.Early Breast Cancer Trialists' Collaborative Group. Tamoxifen for early breast cancer: an overview of the randomised trials. Lancet. 1998;351:1451–1467. [PubMed] [Google Scholar]