Affiliations of authors: M.-H. Shin, Department of Nutrition, Harvard School of Public Health, Boston, MA, and Department of Preventive Medicine, Sungkyunkwan University School of Medicine, Suwon, Korea; M. D. Holmes, Channing Laboratory, Department of Medicine, Brigham and Women's Hospital, Harvard Medical School, Boston; S. E. Hankinson, Channing Laboratory, Department of Medicine, Brigham and Women's Hospital, Harvard Medical School, and Department of Epidemiology, Harvard School of Public Health; K. Wu, Department of Nutrition, Harvard School of Public Health; G. A. Colditz, Channing Laboratory, Department of Medicine, Brigham and Women's Hospital, Harvard Medical School, and Department of Epidemiology, Harvard School of Public Health, and Harvard Center for Cancer Prevention, Boston; W. C. Willett, Department of Nutrition, Harvard School of Public Health, and Channing Laboratory, Department of Medicine, Brigham and Women's Hospital, Harvard Medical School, and Department of Epidemiology, Harvard School of Public Health.
Correspondence to: Walter C. Willett, M.D., Dr.P.H., Department of Nutrition, Harvard School of Public Health, 665 Huntington Ave., Bldg. 2, Boston, MA 02115 (email: wwillett{at}hsph.harvard.edu).
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ABSTRACT |
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INTRODUCTION |
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Components in milk that might be anticarcinogenic include calcium and vitamin D, and some studies have suggested protective effects of calcium and vitamin D against colon cancer (23,24). The hypothesized effect of calcium on colon cancer is intraluminal binding with bile acids and fatty acids, thus reducing the proliferative stimulus of these compounds. With breast cancer, calcium has been proposed to reduce fat-induced cell proliferation by maintaining intracellular calcium concentrations (25,26). Vitamin D modulates calcium metabolism (27) and has calcium-independent antiproliferative actions (26). Some suggest that only vitamin D, not calcium, inhibits mammary tumorigenesis (28), but an independent anticancer effect of higher calcium intake in rats has also been reported (29).
Epidemiologic evidence relating calcium and vitamin D intakes to breast cancer risk is limited. Breast cancer rates in white women are highest in areas with the least winter sunlight and longest winters, which has been interpreted to support an association with vitamin D (30), although we did not find this association in the Nurses' Health Study (NHS) (31). A few epidemiologic studies have reported a statistically significant inverse association between calcium intake and breast cancer (19,3234), while others reported no association (35,36). However, intake of calcium and dairy foods are strongly correlated, so the observed associations with calcium are difficult to separate from those with milk, dairy products and, consequently, other components of milk and dairy products. To our knowledge, there is no epidemiologic study of dietary vitamin D and breast cancer.
We examined data from a large, long-term cohort study to evaluate the hypotheses that higher intakes of dairy products, calcium, or vitamin D are associated with reduced risk of breast cancer. By taking into account supplemental intake of calcium and vitamin D, we hoped to dissociate the effects of calcium and vitamin D from those of milk and dairy products.
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METHODS |
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Semiquantitative Food-Frequency Questionnaires and Calculation of Nutrient Intake
The food-frequency questionnaires have been described in detail, and their validity and reproducibility have been documented previously (37,38). A commonly used portion size was specified for each food (e.g., one 8-oz glass of milk). Participants were asked how often, on average, they had consumed that amount of each food over the past year. The nine prespecified frequency categories ranged from "never" to "6 or more times per day."
The dairy food group included skim/low-fat milk, whole milk, cream, sour cream, sherbet, ice cream, yogurt, cottage cheese, cream cheese, other (hard) cheese, and butter. Total dairy food intake was calculated by summing up the daily servings of all the foods in the dairy group, except butter because it was composed almost entirely of fat and was therefore unlike the other dairy foods. When we calculated dairy fat intake, however, we took into account all the foods in the dairy group, as well as dairy ingredients from other foods. Low-fat dairy food intake was calculated by summing the daily servings of skim/low-fat milk, sherbet, yogurt, and cottage cheese. High-fat dairy food intake was calculated by summing the daily servings of whole milk, cream, sour cream, ice cream, cream cheese, other cheese, and butter. Total milk intake was calculated by summing the daily servings of skim/low-fat milk and whole milk. Total fermented milk intake was calculated by summing the daily intakes of sour cream, yogurt, cottage cheese, cream cheese, and other cheese.
Nutrient intakes were computed by multiplying the frequency of consumption of each unit of food and the nutrient content of the specified portions. Nutrient values in foods were obtained from U.S. Department of Agriculture sources (39). Nutrient intakes were energy-adjusted by using the residuals from the regression of nutrient intake on total caloric intake (40). Residuals were adjusted to 1600 kcal/day, the approximate median caloric intake of all the participants with acceptable diet data, to have meaningful nutrient values.
The information on current use, brand, and dosage of multivitamin supplements was collected on each biennial questionnaire. Current use and dosage of individual calcium and vitamin D supplements were also collected biennially from the 1982 questionnaire.
Total calcium and total vitamin D intakes were calculated from all sources of calcium and vitamin D, combining diet and supplements. Calcium supplement users were defined as users of either individual calcium supplements or multivitamins containing calcium. Vitamin D supplement users were defined as users of either individual vitamin D supplements (tablets) or multivitamins containing vitamin D. We took into account the multivitamin brand and type information supplied by the cohort participants when calculating the supplement intake amount. Dietary calcium and dietary vitamin D intakes were calculated from all dietary sources without supplements. Dietary calcium was further divided according to the food sources, i.e., calcium from dairy sources (dairy calcium) and calcium from nondairy sources (nondairy calcium). Dairy calcium intake was calculated from all dairy foods as well as dairy ingredients from recipes of other foods, for example, cheese from pizza and milk from mashed potatoes. Nondairy calcium was calculated by subtracting dairy calcium from dietary calcium intake. Total nondairy calcium was the combination of nondairy calcium and supplemental intake. We did not divide dietary vitamin D further because the sources are limited to relatively few foods. Calcium supplement users (18.7% of premenopausal and 35% of postmenopausal person-years) were excluded in the primary analyses of dietary calcium and dairy calcium; vitamin D supplement users (31% of premenopausal and 33% of postmenopausal person-years) were excluded in the primary analysis of dietary vitamin D because the supplements could obscure the effects of diet. We also conducted alternative analyses that included the supplement users.
Among 173 members of the NHS cohort, Pearson correlation coefficients (corrected for within-person day-to-day variation) between the means of four 1-week diet records and the 1980 food-frequency questionnaire were 0.69 for skim/low-fat milk and 0.56 for whole milk (38). Correlation coefficients between intakes of calcium from the questionnaire used in 1984 and intakes from four 1-week diet records measured in 1980 among 150 Boston-area women were 0.56 with supplements and 0.51 without supplements (37). Correlation coefficients between intake and plasma concentration of vitamin D in 57 men and 82 women were 0.35 with supplements and 0.25 without supplements (41).
Identification of Breast Cancer Cases
In each biennial questionnaire, participants were asked whether they had been diagnosed as having breast cancer in the previous 2 years. Deaths were identified by a report from a family member, the postal service, or the National Death Index. Medical records were obtained for breast cancer cases identified by either self-report or vital records, and more than 99% of these records confirmed the self-report. Cases of carcinoma in situ were excluded.
Statistical Analysis
Each participant accumulated person-time beginning with the return of the 1980 questionnaire and ending with her cancer diagnosis, death, or May 31, 1996, whichever came first. To avoid recall bias, we started the follow-up of each variable from the time it was collected, even when it was related to previous experience. For example, high school dietary information was collected in 1986, and its analysis included person-time from 1986.
To reduce within-person variation and best represent long-term dietary intake of participants, we assessed the incidence of breast cancer in relation to the cumulative average of dietary intake from all available dietary questionnaires up to the start of each 2-year follow-up interval (42). For example, the incidence of breast cancer from 1980 through 1984 was related to the dietary information from the 1980 questionnaire, and the incidence of breast cancer from 1984 through 1986 was related to the average intake from the 1980 and 1984 questionnaires. When dietary information was missing at the start of the interval, the cumulative average from the previous period was carried forward. To take into account a longer latency of exposure, the 1980 baseline diet also was related to the breast cancer incidence during the entire follow-up period. To consider intermediate latency, we used a 4-year time lag between diet assessment and the start of a follow-up interval. We also analyzed premenopausal diet, that is, the cumulative average diet until menopause, to assess whether postmenopausal breast cancer risk was related to diet in the premenopausal period. A subcohort of premenopausal women at baseline was used for this analysis, and their follow-up was started when they became postmenopausal. To assess the influence of consistency in dairy food intake, we conducted restricted analyses using data from women who had not greatly changed their milk intake during the 10 years prior to baseline in 1980.
All dietary variables were categorized according to quintiles based on their distribution in each questionnaire cycle. Consequently, the cutoff points of the quintiles slightly varied by questionnaire cycles. Categories for dairy foods and nutrients based on their absolute intakes were also created to have common categories throughout the whole follow-up cycle. The relative risks (RRs) and 95% confidence intervals (CIs) were calculated for each category of intake compared with the lowest intake group. We used pooled logistic regression to estimate multivariable RRs using 2-year time increments (43). We simultaneously adjusted for age in 5-year categories, time period, physical activity in metabolic equivalent-hours (METs, with activity at rest = 1.0) values, history of benign breast disease, family history of breast cancer, height, weight change since age 18 years, body mass index (BMI; kg/m2) at age 18 years, age at menarche, parity, age at first birth, alcohol intake, total energy intake, total fat intake, glycemic index value (= [glycemic index x carbohydrate intake for each food]/total carbohydrate intake), -carotene intake, and total vitamin E intake. For postmenopausal women, we added age at menopause and postmenopausal hormone use to the multivariable model. For vitamin D analyses, we added history of outdoor sun exposure and participant's residential area. These covariates were either known or suspected risk factors for breast cancer or had been found to be associated with dairy/calcium/vitamin D intake and with breast cancer risk in the NHS data. RRs were adjusted for glycemic index value because blood glucose level could affect the active cellular uptake of calcium. We calculated two-sided P values for the test for trend by using the Wald statistic. Median values of each intake category were used in the test for trend. Interactions were evaluated by using the likelihood ratio test.
All nondietary covariates included in the multivariable analyses were updated every 2 years, except age at menarche, height, weight at 18 years old, and outdoor sun exposure (yes, if time spent outdoors was 8 hours/day in summer), which were determined at baseline. Activity in METs values was calculated in 1980, 1986, 1988, 1992, and 1994; information on family history of breast cancer (mother, yes/no, or sister, yes/no) was sought in 1976, 1982, 1988, and 1992; and site of residence was recorded in 1976 and updated in 1986, 1988, 1990, 1992, and 1994 and, at each time, was recoded into four regions that included a total of 11 states (California, Northeast, Midwest, and South). Participants who moved out of the 11 original states during the follow-up were categorized as living in an other region. A woman was classified as postmenopausal from the time she returned a questionnaire reporting natural menopause or hysterectomy with bilateral oophorectomy. Women who reported hysterectomy without bilateral oophorectomy were classified as being of uncertain menopausal status until they reached the age at which natural menopause had occurred in 90% of the cohort (54 years for current cigarette smokers and 56 years for nonsmokers), at which time they were classified as postmenopausal. Because we observed differences in the associations of major study variables with breast cancer risk between pre- and postmenopausal women, we report our results from pre- and postmenopausal women separately. Women with uncertain menopausal status were excluded from the analysis.
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RESULTS |
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At baseline, most of the known breast cancer risk factors did not vary appreciably across categories of total milk, total calcium, and total vitamin D intake (Table 1). Calcium, vitamin D, and multivitamin supplement users were more frequent in the high total calcium and vitamin D intake groups. Postmenopausal hormone use and history of osteoporosis were not strongly associated with total milk intake or total calcium intake. Women who consumed more total milk, total calcium, or total vitamin D exercised more, smoked less, and drank less alcohol.
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In postmenopausal women, none of the dairy products had any appreciable association with breast cancer risk (Table 2). However, in premenopausal women, a statistically significant inverse association between low-fat dairy food intake and breast cancer risk was observed (Table 2
). The multivariable RR comparing the highest with the lowest categories of consumption was 0.68 (95% CI = 0.55 to 0.86; Ptrend = .003). Skim/low-fat milk intake accounted for 49% of low-fat dairy food intake in the 1984 data and was the dairy food most strongly related to breast cancer risk (>1 serving/day versus never; RR = 0.72, 95% CI = 0.56 to 0.91; Ptrend = .007). High-fat dairy food and whole milk, which accounted for 12% of high-fat dairy food intake in 1984, showed statistically nonsignificant inverse associations with premenopausal breast cancer risk (for high-fat dairy food, >2.5 servings/day versus
3 servings/week; RR = 0.83, 95% CI = 0.62 to 1.10). When we included the intakes of skim/low-fat milk and whole milk simultaneously in the same model, the RRs were 0.68 (95% CI = 0.53 to 0.88) for skim/low-fat milk and 0.71 (95% CI = 0.49 to 1.05) for whole milk. Total milk intake showed a statistically significant linear inverse association with premenopausal breast cancer (>1 serving/day versus
3 servings/month; RR = 0.69, 95% CI = 0.54 to 0.87; Ptrend<.001).
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There was no qualitative difference between age-adjusted and multivariable RRs. For example, the age-adjusted RR for highest versus lowest intake of low-fat dairy food was 0.77 (95% CI = 0.63 to 0.95) and that for skim/low-fat milk was 0.82 (95% CI = 0.66 to 1.01). Analyses using diet at baseline (1980), diet with a 4-year time lag, and diet among consistent milk consumers yielded RRs that were qualitatively similar but slightly attenuated compared with those in Table 2. Also, premenopausal intakes of dairy foods were not associated with postmenopausal breast cancer risk.
Calcium and Vitamin D
Calcium and vitamin D intakes were not associated with postmenopausal breast cancer risk (Table 3). Dairy and total nondairy calcium also had no association with risk in postmenopausal women. In premenopausal women, most of the calcium- and vitamin D-related variables were inversely associated with breast cancer incidence (Table 3
). The multivariable RR for highest versus lowest total calcium intake was 0.80 (95% CI = 0.58 to 1.12; Ptrend = .05) and that for dietary calcium intake was 0.67 (95% CI = 0.49 to 0.92; Ptrend = .02). When we further divided dietary calcium into dairy and nondairy calcium, dairy calcium was inversely associated with risk (>800 mg/day versus
200 mg/day; RR = 0.69, 95% CI = 0.48 to 0.98; Ptrend = .01); nondairy dietary calcium had no association, although the range in intake was much smaller (>350 mg/day versus
275 mg/day; RR = 1.12, 95% CI = 0.78 to 1.60). Total nondairy calcium also had no appreciable association with risk. Simultaneous inclusion of dairy and nondairy calcium in the model did not change the results appreciably.
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No qualitative difference was observed between age-adjusted and multivariable RRs for the associations of vitamin D and calcium with premenopausal breast cancer risk. The age-adjusted RR for highest versus lowest intake of dietary calcium was 0.74 (95% CI = 0.55 to 0.98) and that for dietary vitamin D was 0.78 (95% CI = 0.52 to 1.18). When we used 1980 baseline intakes of calcium from various sources in the analyses, inverse associations with premenopausal breast cancer were still seen, but the magnitude and statistical significance of trends were attenuated compared with the use of cumulative average intakes (Table 3). Baseline total and dietary vitamin D intakes were not associated with premenopausal breast cancer risk. Similar attenuation was observed when a 4-year time lag was applied between calcium and vitamin D intakes and premenopausal breast cancer. The RR for dairy calcium (>800 mg/day) was 0.78 (95% CI = 0.57 to 1.07; Ptrend = .06). When we included supplement users in the analyses, the associations of calcium and vitamin D intakes with premenopausal breast cancer were slightly attenuated but were not changed substantially. Again, the premenopausal intakes of calcium and vitamin D were not associated with postmenopausal breast cancer risk.
Supplemental calcium had no apparent linear association with breast cancer risk in either premenopausal or postmenopausal women (Tables 4 and 5). In the stratified analysis by tertiles of dietary calcium among premenopausal women, women who used high-dose calcium supplements (
900 mg/day) had a lower risk only if they were in the high dietary calcium intake group (RR = 0.44, 95% CI = 0.19 to 1.01) although the number of case subjects in this analysis was very small. Supplemental vitamin D had a weak, statistically nonsignificant inverse association with premenopausal breast cancer risk, and this association was more prominent among women in the low dietary vitamin D intake group (for
400 IU/day; RR = 0.80, 95% CI = 0.57 to 1.14). Most of the supplemental vitamin D use was determined by multivitamin use. Postmenopausal hormone use did not modify the associations between calcium or vitamin D supplement intake and breast cancer.
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The intake of other constituents of dairy foods had associations similar to those with dairy calcium. Lactose and phosphorus intakes, which were highly correlated with intake of low-fat dairy foods (r = 0.75 and 0.71, respectively), were inversely associated with breast cancer risk in premenopausal women (quintile 5 versus quintile 1; RR = 0.68, 95% CI = 0.54 to 0.86; Ptrend<.001 for lactose and RR = 0.73, 95% CI = 0.56 to 0.96; Ptrend = .01 for phosphorus). Dairy fat intake was associated with lower risk in the highest quintile relative to the lowest, but the trend was not statistically significant (quintile 5 versus quintile 1; RR = 0.78, 95% CI = 0.62 to 0.98; Ptrend = .13).
These inverse associations between the constituents of dairy food intake and premenopausal breast cancer disappeared rapidly after menopause. In early menopause, i.e., within 10 years after menopause, the RRs for dairy calcium, lactose, phosphorus, and dairy fat were already attenuated and statistically nonsignificant. In late menopause, i.e., beyond 10 years after menopause, the RRs were virtually null. For example, for dairy calcium, the RR for the highest versus lowest quintiles was 0.67 (95% CI = 0.52 to 0.88) in premenopausal women, 0.85 (95% CI = 0.65 to 1.11) during early menopause, and 1.05 (95% CI = 0.83 to 1.34) in late menopause. Also, the associations between the constituents of dairy foods with premenopausal breast cancer were more pronounced in women of even younger ages. For example, for dairy calcium (highest versus lowest quintile), the RR of premenopausal breast cancer occurring 6 years or more before menopause was 0.66 (95% CI = 0.38 to 1.12), and within 5 years of menopause, the RR was 0.86 (95% CI = 0.68 to 1.09).
The inverse association between low-fat dairy food intake and premenopausal breast cancer risk was attenuated when it was further adjusted for lactose, dairy calcium, total vitamin D, and phosphorus intakes, but only the adjustment for lactose was strong enough to make the association not statistically significant (highest versus lowest intake of low-fat dairy products; RR = 0.81, 95% CI = 0.61 to 1.08).
Because vitamin D modulates calcium metabolism, effects of dairy calcium might be modified by vitamin D intake. When we stratified the data by tertiles of total vitamin D intake, dairy calcium appeared to be associated with reduced risk of premenopausal breast cancer in women at all levels of vitamin D intake (Pinteraction = .85). According to the calciumhigh fat hypothesis (25,26), the effect of calcium should be stronger in the high-fat-diet group. However, we observed an inverse association between dairy calcium intake and premenopausal breast cancer in all tertiles of total fat intake. Similarly, the association between total vitamin D intake and premenopausal breast cancer was not modified by total fat intake. The inverse association between dairy calcium and breast cancer also was not modified by height, current BMI, or alcohol intake.
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DISCUSSION |
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We hypothesized that if milk intake is associated with reduced risk of breast cancer, calcium or vitamin D might be the responsible nutrient. In the present study, intake of calcium and vitamin D indeed had statistically significant inverse associations with premenopausal breast cancer. Detailed analyses for calcium, however, suggested that the inverse association with calcium was due mostly to dairy sources rather than to nondairy sources or supplements. Low-fat dairy foods remained associated with premenopausal breast cancer risk when total calcium was included in the same model, whereas total calcium was no longer associated with risk (RR for >1250 mg/day versus 500 mg/day of total calcium = 0.95, 95% CI = 0.66 to 1.35). This suggests that dairy calcium is not likely to be responsible for the association of low-fat dairy food and skim/low-fat milk intake with reduced risk of premenopausal breast cancer.
Vitamin D intake, by contrast, was most strongly related to risk when it was analyzed as total intake rather than as dietary or supplemental intake. Unlike calcium, vitamin D intake was largely accounted for by multivitamin supplementation (35% of total intake; 1984 data). Summation of dietary and supplement intake increases the range of intake and, hence, the power to detect an association. Although vitamin D did not contribute substantially to the association between low-fat dairy foods and premenopausal breast cancer risk when both terms were included in a multivariable model, the inverse association with vitamin D was sustained (RR for >500 IU/day versus 150 IU/day of total vitamin D = 0.78, 95% CI = 0.59 to 1.02). Therefore, vitamin D intake appears to offer a possible protective association apart from the "milk effect." Residual confounding by the level of sun exposure could still be possible, because our measures of sun exposure (i.e., outdoor sun exposure and site of residence) were rather crude.
Knekt et al. (19) reported a strong inverse association between milk intake and subsequent incidence of breast cancer in a Finnish cohort study. They also found statistically significant inverse associations of breast cancer with calcium and lactose intakes. Their study was based on only 88 case subjects, but the observed decrease in risk among the highest tertile of total milk intake compared with the lowest was quite strong (RR = 0.42, 95% CI = 0.24 to 0.74). The fact that their cohort included more younger person-years (mean age at baseline = 39 years, with 25 years of follow-up) and consumed more than twice the amount of milk (531 g/day) than was consumed in our cohort (215 g/day) might account for the strength of the association. Two cohort studies conducted in Norway, where people also consume milk frequently (>1.7 glasses/day), reported both positive (17) and negative (20) associations between milk intake and breast cancer. The former study included older person-years (mean age = 43 years, with 10.4 years of follow-up) than the latter study (mean age = 40.7 years, with 6.2 years of follow-up, which was similar to our premenopausal analysis). In addition, the main type of milk consumed in the former study was whole milk, whereas in the latter study, it was skim and low-fat milk. Other cohort studies conducted in the United States, where people consume less than one glass of milk per day, reported either decreased risk with greater intake of milk (18) (age range = 3565 years, with <5 years of follow-up) or no association (16) (median age = 55 years, with 20 years of follow-up). Most of the casecontrol studies with positive associations between milk intake and breast cancer risk also had the common components of older age and consumption of whole milk (12,14, 15,21).
Validity of the dietary measurement should be considered in the interpretation of these conflicting results between studies. However, intake assessments of commonly consumed beverages, such as milk or coffee, have high validity and reproducibility (38). Furthermore, by updating and averaging the repeated measurements of diet five times, we reduced within-person fluctuations and took into account changes over time, which should improve validity compared with most previous studies.
Confounding might be another reason for the conflicting results in previous studies. In the present study, we adjusted all food and nutrient items for total energy intake, so that the relative rather than the absolute amount of intake could be evaluated (40). We also controlled for a wide range of potential confounders. Adjustment for other nutrient factors changed the result meaningfully in this study. For example, the RR of premenopausal breast cancer for the highest intake of total milk relative to the lowest was 0.78 (95% CI = 0.64 to 0.96) in the age-adjusted model, 0.77 (95% CI = 0.62 to 0.95) in a multivariable model without other dietary factors, and 0.69 (95% CI = 0.54 to 0.87) in the fully adjusted model. Changes in the RRs for calcium and vitamin D were even greater than those for total milk.
The apparent association of intake of milk and its related nutrients with reduced risk of breast cancer disappeared quickly after menopause. This diminishing association with increasing age was also observed in an Italian casecontrol study (45). It is not clear why milk and its related nutrients might show different associations with pre- and postmenopausal breast cancer.
Components of milk other than vitamin D and calcium might be responsible for the associations we observed. Lactose can aid the absorption of dietary calcium as well as promote the growth of lactic acid-producing bacteria in the large intestine. Although lactose intake is suspected to be associated with an increased risk of ovarian cancer (46) and decreased risk of colon cancer (47), little is known about its relationship with breast cancer. Lactose has been hypothesized to increase ovarian cancer risk by direct toxicity to oocytes and by inducing premature ovarian failure (48). This effect could reduce exposure of breast tissue to estrogen. Some epidemiologic studies reported an inverse association between fermented milk (6) and yogurt (12) and risk of breast cancer, interpreting this as an effect of lactose or lactic acid.
Other components of milk have the potential to explain the apparent protective association with breast cancer. Conjugated linoleic acid (CLA), a mixture of positional and geometric isomers of linoleic acid, comes from dairy (60%) and beef (32%) products (49) and is a potent anticarcinogen in animal models (50). However, the inverse association we saw with dairy foods was probably not due to CLA because it was strongest for low-fat dairy products, which should have low CLA content.
In conclusion, high intakes of dairy products, especially low-fat dairy and skim/low-fat milk, may be associated with a modest reduction in the risk of breast cancer in premenopausal women but not in postmenopausal women. Other constituents of dairy foods, such as calcium, total vitamin D, lactose, and phosphorus, also showed inverse associations with risk of premenopausal breast cancer, but their independent associations with breast cancer are difficult to distinguish. Further study of the relationship between dairy product consumption and breast cancer is warranted, with a specific focus on premenopausal women.
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NOTES |
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We thank Karen Corsano, Diane Feskanich, Laura Sampson, Debbie Flynn, and Francine Laden for their technical support and valuable advice.
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Manuscript received January 16, 2002; revised June 3, 2002; accepted July 17, 2002.
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