1 Division of Cancer Prevention and Control, Centers for Disease Control and Prevention, Atlanta, GA.
2 Cancer Prevention, Detection, and Control Research Program, Duke University Medical Center, Durham, NC.
3 Department of Epidemiology, School of Public Health, and Lineberger Comprehensive Cancer Center, School of Medicine, University of North Carolina, Chapel Hill, NC.
4 School of Public Health, Queensland University of Technology, Kelvin Grove, Queensland, Australia.
Received for publication December 9, 2003; accepted for publication June 24, 2004.
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ABSTRACT |
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African Americans; breast neoplasms; case-control studies; risk factors; women
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INTRODUCTION |
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Distinguishing breast cancers by age at onset has important implications for cancer incidence and etiology. The most recent Surveillance, Epidemiology, and End Results Program statistics for 19962000 found the age-adjusted incidence of breast cancer to be higher among White than African-American women (142.0 per 100,000 vs. 120.8 per 100,000) (7). The same pattern was seen in a comparison of White and African-American women aged 50 years or older (398.6 vs. 322.1 per 100,000), but the trend was reversed among women aged less than 40 years. In that group, the incidence rate for each 5-year age group was higher among African-American women (7). This disparity indicates that African-American women are experiencing excess rates of breast cancer at younger ages. Evaluation of age distributions in case series has shown that African-American breast cancer patients are more likely to present at a younger age (811) and that cases of breast cancer among young women overall are more aggressive and show a poorer prognosis and response to treatment than cases among older women (8, 1115). These factors contribute to a mortality rate for breast cancer among younger African-American women that is twice that of younger White women (7).
To evaluate potential differences in risk factors for breast cancer, we compared risk factor profiles between African-American women and White women, stratified at age 50 years, in the Carolina Breast Cancer Study. To highlight differences in risk factors that may contribute to excess incidence among younger African-American women, we first present adjusted odds ratios for reproductive and lifestyle factors. Because breast tumor characteristics are known to differ between African-American and White women (16), we then analyzed the pattern of results from case-only analyses before and after further adjustment for stage and hormone receptor status.
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MATERIALS AND METHODS |
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Controls were drawn from North Carolina Division of Motor Vehicles lists for women aged 2064 years and US Health Care Financing Administration lists for women aged 6574 years. Sampling probabilities for controls ensured approximate frequency matching to cases by race and 5-year age groups. Of the 2,444 eligible and locatable women, 727 (30 percent) declined to participate. Thus, information from 1,717 controls was available for analysis, for an overall cooperation rate of 70 percent. Cooperation rates for the four age-race groups were 76 percent, 71 percent, 72 percent, and 63 percent for younger White and African-American controls and older White and African-American controls, respectively.
After exclusion of 276 women who agreed only to a brief telephone survey and had incomplete information on many risk factors of interest, the final data set consisted of 1,505 (45 percent) African-American women (787 cases and 718 controls) and 1,809 (55 percent) White women (991 cases and 818 controls).
Data collection
The data were obtained during in-person interviews conducted by female registered nurses. Through a questionnaire, the nurse-interviewer elicited information on demographics and potential breast cancer risk factors, including first-degree family history of breast and ovarian cancer, menstrual and reproductive history, and sociodemographic and lifestyle characteristics. The nurses drew a blood sample and measured weight, height, and waist and hip circumferences at the time of interview. The median time from diagnosis to interview for the cases was 3 months (range: 119 months); 80 percent were interviewed within 5 months of diagnosis. For the controls, the median time from selection to interview was 2 months (range: 026 months); 80 percent were interviewed within 5 months of selection.
The American Joint Committee on Cancer stage was abstracted from medical records, where available, or determined from information on tumor size, lymph node involvement, and distant metastasis (16). Estrogen receptor and progesterone receptor status was obtained from medical records for 80 percent of the cases. For the remaining cases, estrogen receptor and progesterone receptor status was determined with paraffin-embedded tumor tissues at the University of North Carolina laboratory (11 percent), or receptor status was missing (9 percent) (19, 20).
Statistical methods
Statistical analyses were performed separately by race, and they were further stratified by age or menopausal status. Women were categorized as postmenopausal if they reported natural menopause or bilateral oophorectomy or if they were more than 55 years of age and reported hysterectomy. Women who reported still having menstrual cycles or who had at least one remaining ovary and were aged less than 42 years were classified as premenopausal. Women who had a hysterectomy without bilateral oophorectomy and were aged 4255 years were considered perimenopausal.
The frequency distributions of risk factors in the study sample were adjusted using age-specific sampling weights to estimate prevalences in the underlying population. Comparisons by race were evaluated using the chi-square statistic. Both the prevalence estimates and chi-square tests were generated using SUDAAN version 8.0.0 software (Research Triangle Institute, Research Triangle Park, North Carolina). Odds ratios with 95 percent confidence intervals were calculated using logistic regression models that examined the association between breast cancer status and risk factors after adjustment for other relevant covariates including age (continuous), age at menarche, parity, age at first full-term pregnancy, miscarriage, breastfeeding, induced abortion, oral contraceptive use, and hormone replacement therapy. Body mass index, waist/hip ratio, history of breast cancer in a first-degree relative, education, alcohol consumption, and smoking were also included as covariates in addition to a term for the sampling fraction. Analyses also were adjusted for years since last full-term pregnancy; these results are not presented because odds ratios were equivalent within plus or minus 0.1. Logistic analyses were performed using SAS PROC GENMOD software (SAS Institute, Inc., Cary, North Carolina), which permits the use of an offset term to take the sampling design into account.
Tests for interaction between each covariate and race were conducted by performing the likelihood ratio test using data with both races combined. A model containing only the main effects of race, a variable of interest, and other covariates was compared with a model containing the main effects of race, the variable of interest, the relevant interaction term, and other covariates. Tests were conducted for interaction among all women, younger women, and older women. None of the interactions by race reached statistical significance at p < 0.05; however, a few comparisons yielded likelihood ratio tests with p < 0.20 (noted in text). Because of the limited statistical power for tests of interaction, we also chose to highlight differences between risks for African-American women and White women of 40 percent or greater in magnitude of the odds ratio or odds ratios that went in opposite directions. These criteria, though admittedly arbitrary, were selected to avoid making too much of fairly small variations in odds ratios and to focus attention on variations in odds ratios by race that were potentially meaningful in our data. We thought such results warranted further examination in the discussion in relation to possible consistency with other available findings.
To adjust for racial differences in breast cancer characteristics (i.e., stage at diagnosis and hormone receptor status), we conducted a series of case-only analyses using logistic regression models. Odds ratios and 95 percent confidence intervals were derived from case-case comparisons to highlight the presence of heterogeneity between African-American and White women with breast cancer after adjustment for stage and estrogen receptor and progesterone receptor status (i.e., odds ratios deviating from 1.0 suggest racial differences). Although the case-case odds ratio can be used as a measure of heterogeneity of odds ratios (21), its magnitude reflects risk factor differences in both background prevalence and disease-associated risk, and the absence of noncase comparisons limits etiologic inferences. However, comparison of unadjusted and adjusted case-only odds ratios can provide an indication as to whether the tumor characteristics are potentially important to racial differences in risk factors.
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RESULTS |
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In table 2, associations between breast cancer and various reproductive and lifestyle characteristics are shown for younger women by race; differences in odds ratios by race were observed for most reproductive characteristics and also former smoking. Younger African-American women with average age at menarche (1213 years) were at increased risk of breast cancer, as were their White counterparts (but not to the same degree). For younger White women, the risk increased with later age at first pregnancy, and nulliparous women were at greater risk than were parous women. In contrast, among African-American women, later age at first pregnancy did not increase risk, and nulliparous African Americans were at slightly reduced risk versus those who had a first pregnancy before age 25 years and those with three or more children. African-American women who breastfed were at reduced risk of breast cancer, but White women were not (pinteraction < 0.12). An interaction with race was also suggested for age at menopause, with odds ratios farther from 1.0 at both younger (aged 44 years) and later (aged 4549 years) ages among younger White women than younger African-American women (pinteraction < 0.19). In addition, former African-American smokers were at elevated risk, but former White smokers were not (pinteraction < 0.06). Patterns and magnitudes of risk were similar by race for family history of breast cancer, body size, waist/hip ratio, education, oral contraceptive and hormone replacement therapy use, alcohol consumption, and induced and spontaneous abortion.
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DISCUSSION |
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Several reproductive risk factors were among those with apparent variations in prevalence and patterns of breast cancer risk between African Americans and Whites. Older age at first full-term pregnancy was associated with a modest increased risk among White women, with nulliparous women being at highest risk, as is traditionally reported. This was evident for both younger and older women among Whites; however, the pattern did not hold for African-American women. Among younger African Americans, no associations with older age at first full-term pregnancy or nulliparity were observed. Among older African-American women, increasing age at first full-term pregnancy showed no association with breast cancer, but nulliparous women were at twice the risk of parous women. Palmer et al. (25) also found no elevation of risk due to late age at first birth in older African-American women. Brinton et al. (6) obtained similar results, reporting that nulliparous younger African-American women were at reduced risk of breast cancer relative to multiparous women. In addition, Laing et al. (4) reported no increased breast cancer risk for nulliparous African-American women compared with multiparous African-American women. Hence, it appears that neither age at first full-term pregnancy nor nulliparity helps to explain the observed racial pattern of breast cancer incidence.
Multiparity is often reported to reduce breast cancer risk in comparison with women having no children (26). We observed this relation among both younger and older White women, although there was no evidence of a dose-response trend. Again, however, relations among African-American women tended to vary by age. While a similar inverse relation was observed for older African Americans, younger African-American women with three or more children had an odds ratio 4050 percent higher (not statistically significant, however) than the odds ratio for their nulliparous peers. They also were almost twice as likely to have families of this size as younger White women. These results replicate the findings of Palmer et al. (25), who documented that parity is associated with increased risk in younger African-American women but with decreased risk among older women. An age-dependent effect of increasing parity on breast cancer risk has been reported previously (2729). In one study, Bruzzi et al. (27) provided evidence for a positive association between breast cancer and increasing parity among women younger than age 40 years with two or more children, but (as with our data for African Americans) the trend was not significant, while they reported an inverse relation for older women. These authors also reported a transient increase in breast cancer risk following full-term pregnancy; a relative risk of 2.66 (95 percent CI: 1.31, 5.39) was obtained for women having delivered a child within 3 years versus women whose last child was born 10 or more years earlier. If younger African-American women have children at intervals of 3 years or less, the transient risk period could be prolonged. In addition, if having multiple children actually increases risk of breast cancer for young African-American women, the higher prevalence of this factor among African Americans could serve to elevate risk for breast cancer in this population, consistent with the higher incidence of breast cancer reported for younger African-American women.
If, as a consequence of higher parity, young African-American women were more likely to have a recent pregnancy (10 years previously) than were young White women, the relevant exposure may be time since last full-term pregnancy. These age-specific differences in odds ratios between African-American and White women might be expected if age since last full-term pregnancy is the relevant exposure and if younger African-American women were more likely to have a recent pregnancy. However, among both younger and older parous women in our study, there were no racial differences in distribution of time since last full-term pregnancy. Additionally, when the variable years since last full-term pregnancy was substituted for age at first full-term pregnancy, we saw the expected trend of an increased risk followed by a progressive reduction in odds ratios for younger White women, but no association was seen for younger African-American women (data not shown). Moreover, the odds ratios for parity were virtually unchanged after adjustment (data not shown). We had insufficient variability in time since last full-term pregnancy to conduct similar analyses for older women. Nevertheless, these results provide additional evidence for variation by race in relations between breast cancer and characteristics of reproductive history among younger women.
Previous results from the Carolina Breast Cancer Study on the effect of breastfeeding on breast cancer risk showed a 30 percent reduction for parous women who had ever breastfed relative to parous women who had not (30). Our race-specific analyses showed younger African-American women to be at greater reduced risk (OR = 0.6, 95 percent CI: 0.4, 0.8) than their White peers (OR = 1.0, 95 percent CI: 0.7, 1.4). The inverse association also was stronger among younger African-American women than their older counterparts (OR = 0.6 and OR = 0.8 for younger and older, respectively). We noted that, among older women, the prevalence of breastfeeding was higher among African-American than White women. In contrast, only 20 percent of younger African-American women, almost half the proportion of younger White women, had breastfed. Thus, many younger African-American women may not be benefiting from the potential protective effects of lactation.
Selection bias may have influenced our results, as response rates varied across age, race, and case status (31). Women who declined an in-person interview were asked to complete a brief telephone survey on basic breast cancer risk factors. Both cases and controls who responded only to the telephone survey were older, had an earlier age at first full-term pregnancy, and had less education, oral contraceptive use, and hormone replacement therapy use than women who participated fully. However, the differences were limited and in the same direction for cases and controls, minimizing the concern about selection bias (31). Moreover, comparisons by race for prevalence of risk factors when based only on women with complete data provide conservative estimates of differences. A further refinement took into account differences in disease characteristics between African Americans and Whites in supplementary case-only analyses. Adjustment for stage at disease diagnosis and hormone receptor status of the breast cancer failed to reduce racial differences in risk factor profiles among cases, suggesting that any differences in etiologic pathways are unlikely to be explained by stage at diagnosis or estrogen and progesterone receptor status alone.
Even with the relatively large numbers of cases and controls in this study, the sample sizes were modest for many comparisons when analyses were stratified by race and age; hence, confidence intervals were broad and firm conclusions not possible. Although the findings we highlighted did not represent statistically significant differences by race, we thought it was important to focus on point estimates to discern whether any patterns may emerge. This was particularly of interest because we had roughly comparable numbers of African-American and White women from the same geographic area. With the increasing availability of studies that have included reasonable numbers of African-American women (16, 25), a meta-analysis of results may provide more meaningful estimates of risk factor profiles by race.
In conclusion, published reports on the etiology of breast cancer among African-American women are sparse and often conclude that their risk factors are similar to those for White women. Our results, however, show racial differences in the prevalence of most risk factors. Furthermore, the associations between breast cancer and some of these factors appear to vary in magnitude and direction by race and age. Most of the variations in risk factors occurred among women under 50 years of age, and the observed differences cannot be explained by either stage at diagnosis or hormone receptor status. Our results therefore demonstrate the importance of assessing modification by age when discerning risk estimates, and they support the hypothesis that racial variations in risk combined with racial differences in prevalence for particular risk factors may contribute to the higher incidence of breast cancer among younger African-American women.
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ACKNOWLEDGMENTS |
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The authors thank the nurses of the Carolina Breast Cancer Study for their diligence in collecting questionnaire and medical record data and body measurements and Dr. Wen-Yi Huang and Dr. Lynn Dressler for analysis of the estrogen and progesterone receptor status on archived tumor tissue. A special acknowledgment is warranted for Jessica Tse for guidance and assistance with statistical analysis.
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NOTES |
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REFERENCES |
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