1 University of Minnesota, Minneapolis, MN.
2 University of Utah, Salt Lake City, UT.
Received for publication December 31, 2003; accepted for publication May 11, 2004.
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ABSTRACT |
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age factors; breast neoplasms; female; risk factors
Abbreviations: Abbreviations: CI, confidence interval; IWHS, Iowa Womens Health Study.
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INTRODUCTION |
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The incidence of breast cancer peaks at the ages of 7579 years, and among women diagnosed with breast cancer during 19962000, 44.2 percent were aged 65 or more years at diagnosis, and 22.5 percent were aged 75 or more years (4). Although some studies have reported on risk factors for breast cancer by age at diagnosis, and heterogeneity of breast cancer risk factors between pre- and postmenopausal women is recognized, data describing risk factors for breast cancer in the oldest age groups are relatively limited. For example, in a report based on pooled data from multiple studies (5), only about 7 percent of cases in the data set were aged 70 or more years at diagnosis. Among other studies that have considered the age-specific role of breast cancer risk factors in postmenopausal women (614), most have treated all women aged 65 or more years as a single age group.
Differences in the pathologic features of breast tumors in aging women have been described. The proportion of women with distant-stage disease increases with age (15, 16), the distribution of histologic types changes (15), and the proportion of tumors expressing hormone receptors (17) and other favorable biologic markers (18) increases. These trends appear to continue beyond age 70 years. Differences in pathology may reflect different biologic influences on breast tumor development in older women.
We have analyzed data from a cohort in which a significant number of women have been diagnosed with breast cancer at age 75 or more years. The goal of this analysis was to describe the age-specific influence of recognized risk factors for breast cancer in aging women.
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MATERIALS AND METHODS |
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Forty-three percent of women returned the questionnaire. Women were excluded who at baseline reported a history of cancer other than skin cancer (n = 3,830), a prior mastectomy (n = 547), or premenopausal status (n = 354). Additionally, women who did not answer questions addressing the risk factors of interest for our current analysis were excluded (n = 447). After exclusions, there were 36,658 women, 99 percent of whom reported race as White, with a median age of 61 years at baseline, in the eligible cohort.
Incident breast cancer cases (International Classification of Diseases code 174 and code C50 from the Ninth Revision and Tenth Revision, respectively) through 2001 were identified through annual linkage to the State Health Registry of Iowa, which is part of the National Cancer Institutes Surveillance, Epidemiology, and End Results Program (4). There was virtually no difference between the rates of breast cancer among the responders and nonresponders to the baseline questionnaire (22). Person-years of follow-up were calculated from the date of the baseline questionnaire until the date of breast cancer diagnosis, date of move from Iowa, or date of death. If none of these events occurred, the end of follow-up was December 31, 2001, when the median age of 29,687 surviving participants was 77 years.
Data analysis
The primary exposures of interest for the present analysis are recognized risk factors for postmenopausal breast cancer, including anthropometric measurements, reproductive events (age at first livebirth and parity), variables representing exposures to endogenous hormones (age at menarche, age at menopause), and family history of breast cancer. From the self-reported anthropometric data, we calculated body mass index (weight (kg)/height (m)2), waist/hip ratio (an indicator of body fat distribution), and weight change from age 18 years to baseline, and we established cutpoints based on the distribution in all eligible cohort members. For family history of breast cancer, we made comparisons of 1) women who reported first-degree relatives (mothers, sisters, and daughters) diagnosed with breast cancer with those who reported no family history and 2) women who reported second-degree relatives diagnosed with breast cancer (grandmothers and aunts), but no first-degree relatives, with the same referent group.
To analyze associations between risk factors and incident breast cancer, we used the Cox proportional hazards regression method with age as the time scale. To control for potential confounders of the variables of interest, we selected covariates considered to be risk factors according to previous breast cancer studies, including previous IWHS analyses. Multivariate regression analyses were conducted with the following covariates: age at baseline (continuous), education (less than high school, high school, more than high school), age in years at first livebirth (<20, 2024, 2529, 30, nulliparous), parity (12, 34,
5, nulliparous), age in years at menarche (
11, 12, 13,
14), age in years at menopause (<45, 4549, 5054,
55), family history of breast cancer (none, second-degree relative but no first, first-degree relatives), and body mass index and height (quartiles).
Hazard ratios and 95 percent confidence intervals were calculated for each of three age intervals: 5564 years, 6574 years, and 7584 years. Each woman contributed person-years at risk to one or more of these age categories, depending on her age at entry into the study and her age at end of follow-up. Trends in hazard ratios across ordered exposure categories (e.g., quartiles of body mass index) were examined using a continuous variable (e.g., coded 0, 1, 2, 3) as the explanatory variable. Because the cutpoints for 10-year categories may be arbitrary, we constructed graphs of hazard ratios across age. For these graphs, we calculated a point estimate of the hazard ratio for breast cancer at each single year of age from Cox models that included time-dependent covariates. The point estimates were plotted versus age on a log scale. A line representing the trend in hazard ratio with age was obtained by linear regression of the single-year estimates across age, weighting each estimate according to the number of incident breast cancers at that age.
We formally assessed differences in hazard ratios across the three categories of age at diagnosis by comparing two models. The first model included time-dependent covariates for the risk factor of interest, allowing a different hazard ratio to be estimated for each of three age intervals: 5564, 6574, and 7584 years. The second model did not include time-dependent covariates, constraining the hazard ratio to be constant across all age intervals. Interaction between risk factors and age was tested by comparing the two models using the likelihood ratio test. Stata statistical software (Stata Corporation, College Station, Texas) was used for data analysis.
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RESULTS |
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Characteristics of breast cancer cases by age at diagnosis are shown in table 1. Breast cancer incidence was highest in women aged 7579 years. The majority of tumors were local stage at diagnosis (63 percent of tumors with stage information), with the proportion that was local generally increasing with age. The proportion of tumors positive for the estrogen receptor (84 percent of all those with estrogen receptor results) also increased with age. The proportion with ductal histology decreased with age at diagnosis, while the proportion with histologies classified as "other" (including mucinous adenocarcinoma, tubular adenocarcinoma, medullary carcinoma) was highest in women aged 75 or more years.
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Menarche at older than 11 years was associated with a decreased risk of breast cancer at age 5564 years (table 3), although there was not a clear trend across the categories of women who reported menarche at the ages of 12, 13, and 14 or more years. The hazard ratios for older age at menarche were attenuated for breast cancer diagnosed at ages 6574 and 7584 years compared with age 5564 years. Older age at menopause was modestly associated with increased risk of breast cancer. When the contrast was simplified to compare two categories of age at menopause, hazard ratios for menopause at age 50 or more years, compared with that at age 49 or less years, were 1.11 (95 percent CI: 0.92, 1.35) for age 5564 years at diagnosis, 1.05 (95 percent CI: 0.94, 1.17) for age 6574 years at diagnosis, and 1.26 (95 percent CI: 1.06, 1.49) for age 7584 years at diagnosis. The graph of hazard ratios for age at menopause of 50 or more years versus less than 50 years (figure 2, part C) has a slight positive slope, but the test for interaction did not indicate an important difference in the hazard ratio for age at menopause by age at diagnosis (p = 0.60).
In the present data, there was little relation between family history of breast cancer and breast cancer risk at age 5564 years, but increased risk associated with family history was present for ages 6574 years and 7584 years. The graph of hazard ratios for first-degree family history of breast cancer at each year of age had a positive slope with age (figure 2, part D). The test for interaction provided limited evidence (p = 0.10) of heterogeneity of the hazard ratios for a first-degree family history of breast cancer among the three age groups.
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DISCUSSION |
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Breast cancer incidence in this cohort followed the age trend reported in national data (4), with increasing incidence in each 5-year age group up to 7579 years and then somewhat lower incidence at age 8084 years. Overall similarity of breast cancer incidence between IWHS participants and the population from which they were drawn has been described previously (22). Breast cancers occurring in women who were aged 75 or more years at diagnosis were characterized by a somewhat higher proportion with three favorable prognostic characteristics: local stage at diagnosis, positive estrogen receptor status, and "other" histologies (tumors that were neither ductal nor lobular). The patterns for estrogen receptor status and histology are in agreement with what has previously been described (15, 17, 18). In Surveillance, Epidemiology, and End Results Program data from the 1980s (15) and from 1992 (16), women in the oldest age groups more often had advanced stage breast cancer. This was not true in the more recently diagnosed IWHS cases, a difference that is probably explained by the continued increase in use of mammography screening during the last decade by women aged 65 or more years (23).
The strengths of this study were that it was population based and prospective, with the same participants being followed continuously over a period of 16 years for breast cancer incidence. Because the cohort was followed from a median age of 61 years to a median age of 77 years, most cohort members contributed person-time in all three age categories that were considered in the analysis. To further control for any possible birth cohort effects, analyses were adjusted for age at baseline. Generalizability is a potential limitation; IWHS participants live in one state, are almost exclusively Caucasian, had to have a drivers license in 1986, and had to be able to respond to a detailed written questionnaire. Thus, these women differ from the general population of US women at risk for breast cancer in racial and geographic distribution and literacy. For this analysis, we have emphasized comparisons of the disease experience of women within the cohort at different ages. The results therefore should be internally valid. The time between measurement of exposures at baseline and occurrence of breast cancer varied widely. When baseline responses are used to assess hazard ratios for disease events through a long follow-up period, accuracy of exposure information may diminish with time, for example, if subjects change their dietary habits from what was reported at baseline. There is the potential for hazard ratios for events occurring later during follow-up to be attenuated because of misclassification of exposure. For most of the factors that we emphasized in this analysis, however, for example, parity, age at menarche, and age at menopause, the exposure could not change after baseline among these postmenopausal women, so any attenuation of hazard ratios with age would not be explained by this source of measurement error.
In the existing literature on breast cancer risk factors, age heterogeneity is most evident when the role of body fatness, usually measured by body mass index, is considered. Many studies have shown that the relation between high body mass index and breast cancer is null or slightly inverse when cancers diagnosed before menopause are considered, but high body mass index is associated with an increased risk of breast cancer for postmenopausal women (5, 11, 24). Among prior studies that considered the role of high body mass index by age group among postmenopausal women, most have suggested that relative risks were stronger in the oldest age groups (5, 1113), although one reported the reverse (25). For example, in one study the odds ratio for the highest versus the lowest quintile of body mass index increased from 1.5 for women diagnosed at age 6069 years to 2.9 among women aged 70 or more years at diagnosis (11). Those analyses were based on relatively small numbers of cases in the oldest age groups. In the IWHS cohort, based on 1,297 cases diagnosed at age 6574 years and 561 cases diagnosed at age 75 or more years, we observed that the magnitude of the hazard ratio for body mass index in postmenopausal breast cancer was similar across all ages at diagnosis. Several measures of body fatness (body mass index, waist/hip ratio, and weight change since age 18 years) each had a strong positive association with breast cancer risk for women in all three age groups that we considered. Thus, it appears that women who heed public health messages to maintain a healthy weight in adulthood will have lower breast cancer risk in their elderly years.
An early age at first livebirth and a higher number of births are recognized as protective against postmenopausal breast cancer (26). The present data provide evidence that, among parous women, a high number of births (5 births) is associated with decreased breast cancer risk in all age groups of postmenopausal women, including those aged 75 or more years. We found that nulliparity and age at first livebirth had little association with breast cancer in those aged 75 or more years. Several other studies that reported on parity, age at first birth, and breast cancer by age at diagnosis (69) found higher risk ratios overall for nulliparous women than did the present study, and these studies did not find diminished risk ratios among elderly women. The inconsistency between the result for women aged 75 or more years in the present study and these reports may be explained by chance. There is also the possibility that categorizing women aged 75 or more years as a separate group, rather than grouping all women aged 65 or more years (79) or 70 or more years (6), may be revealing different information.
Older age at menarche typically is reported to be associated with reduced breast cancer risk, while older age at menopause is associated with increased risk (26). In our analysis, the association with age at menarche was present only in the group aged 5564 years, with little evidence of an effect of age at menarche on breast cancer risk in women aged 65 or more years. In the present data, the association between older age at menopause and breast cancer risk was not strong but was present in women aged 75 or more years. These results are consistent with an analysis based on case-control study data, which also indicated that age at menarche was no longer relevant to breast cancer risk after the age of 64 years, but age at menopause continued to influence risk (10).
A pooled analysis of data from population-based studies reported on the association between family history of breast cancer and breast cancer risk by 5-year age categories (27). The risk ratio for having one affected relative compared with none decreased from 2.53 among women aged 3539 years, to 1.84 for women aged 4549 years, to 1.53 for women aged 5559 years, consistent with the idea that inheritance of a high-risk cancer susceptibility gene is associated with both an increased risk of cancer and a tendency toward earlier age at onset. In the same study, risk ratios did not further diminish in older age groups; for women aged 70 or more years, the risk ratio was 1.64. Among the postmenopausal women in the IWHS, the hazard ratio for a first-degree family history of breast cancer was higher in the groups aged 6574 and 7584 years than in the group aged 5564 years. The results of the present study within a single cohort observed over 16 years are in agreement with the pooled analysis of multiple studies (27) in indicating that family history of breast cancer, and by implication genetic susceptibility, continues to predict elevated breast cancer risk in elderly women.
Epidemiologic theory predicts that risk ratios will be stronger in populations with lower absolute risk (28). An example of this phenomenon can be drawn from cardiovascular disease, for which recognized risk factors, for example, smoking or high blood pressure, show larger risk ratios for younger, low-risk subjects than for older, high-risk subjects within the same cohorts (29). The findings of lack of association of nulliparity, age at first livebirth, and age at menarche with breast cancer at age 75 or more years in the IWHS are consistent with attenuation of risk ratios as breast cancer incidence rises with age. However, for several other breast cancer risk factors, we did not observe lower risk ratios for women aged 75 or more years. The risk factors that showed attenuated risk ratios, that is, age at menarche, nulliparity, and age at first birth, represent hormonal exposures that occurred in the distant past. Indicators of relatively recent exposure to endogenous hormones (postmenopausal obesity and age at menopause) appear to remain relevant for breast cancer risk in elderly women.
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ACKNOWLEDGMENTS |
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The authors thank Ching-Ping Hong, Dr. David R. Jacobs, and Peter Hannan for useful suggestions regarding the data analysis.
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NOTES |
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REFERENCES |
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