Recent studies have identified several single-nucleotide polymorphisms (SNPs) associated with the risk of breast cancer and parity and age at first childbirth are well established and important risk factors for breast cancer. The aim of the present study was to examine the interaction between these environmental factors and genetic variants on breast cancer risk.The Malmö Diet and Cancer Study (MDCS) included 17 035 female participants, from which 728 incident breast cancer cases were matched to 1448 controls. The associations between 14 SNPs and breast cancer risk were investigated in different strata of parity and age at first childbirth. A logistic regression analysis for the per allele risk, adjusted for potential confounders yielded odds ratios (OR) with 95% confidence intervals (CI).Six of the previously identified SNPs showed a statistically significant association with breast cancer risk: rs2981582 (FGFR2), rs3803662 (TNRC9), rs12443621 (TNRC9), rs889312 (MAP3K1), rs3817198 (LSP1) and rs2107425 (H19). We could not find any statistically significant interaction between the effects of tested SNPs and parity/age at first childbirth on breast cancer risk after adjusting for multiple comparisons.The results of this study are in agreement with previous studies of null interactions between tested SNPs and parity/age at first childbirth with regard to breast cancer risk. The risk of breast cancer among first-degree relatives of a breast cancer patient is about twice as high as in the general population . The genGenome wide association studies (GWAS) have recently identified several single nucleotide polymorphisms (SNPs) associated with breast cancer risk -7. CertaThe Malmö Diet and Cancer Study (MDCS) is a prospective population-based cohort in Malmö, Sweden. It provides tumour endpoints, DNA samples, and information on parity and other reproductive factors for a total of 17 035 women. The aim of the present nested-case control study was to study whether the reported associations between reported risk SNPs and breast cancer differ by parity and age at first birth.2[The MDCS, a population-based prospective cohort study recruited participants between 1991 and 1996. All female residents of Malmö, Sweden, born between 1923 and 1950 were invited. Written informed consent was obtained by all participants at baseline. In all, 41% of invited women participated, and the female cohort consisted of 17 035 women . BaselinThe MDCS and the present analyses were approved by the Ethical Committee at Lund University .Information on parity was assessed from the questionnaire. The question: “How many children have you given birth to and in what years were they born?” was categorised as; nullipara, one child, two children and three or more children. Parity was further dichotomized as nulliparous and parous in order to yield larger groups. Age at first childbirth was calculated from the information provided in the same question and categorised as; ≤20, >20 - ≤25, >25 - ≤30 and >30. Age at first childbirth was also dichotomized as ≤ 25 years of age and > 25 years of age. Information on parity and age at first childbirth was missing for a small number of women, and they were excluded from all analyses. No reliable information on twin pregnancies was available nor on miscarriages or abortions.All women were followed until 31st of December 2007. Tumour end-points were retrieved by record linkage with The Swedish Cancer Registry (until 31st of December 2006), and due to a delay in central registration, also with linkage to its regional branch, The Southern Swedish Regional Tumour Registry concerning 2007. Vital status was obtained from The Swedish Cause-of-Death Registry until 31st of December 2007.All 17035 women were followed for cancer as described above. For the purpose of the present study, only women with no previous  cancer (not including cancer in situ of the uterine cervix) were eligible for inclusion. A total of 545 cases with incident breast cancer were identified in a first set with follow-up until 31st of December 2004. One case did not have any DNA sample hence this case was excluded. The remaining 544 cases were matched to two controls each, a total of 1088. The matching criteria were age (+/− 90 days) and time of sampling at baseline (+/− 30 days). A new linkage was performed with follow-up until 31st of December 2007, where an additional 186 cases and 372 controls were identified. A total of 11 controls from the first set were diagnoses with breast cancer during the second follow-up period. They were removed as controls and replaced by other controls matched on the same criteria. For 14 women, there were no DNA available for sampling (2 cases and 12 controls); hence they were excluded from all statistical analyses. Finally, the study population consisted of a total of 2176 women out of which 728 were cases and 1448 were controls.Eleven SNPs were selected from previous GWAS and candidate SNP publications by Easton et al. (rs2981582 (FGFR2), rs3803662 (TNRC9), rs12443621 (TNRC9), rs98051542 (TNRC), rs889312 (MAP3K1), rs3817198 (LSP1), rs2107425 (H19), rs13281615 (8q24), rs981782 (5p12), rs30099 (5q), rs4666451 (2p)) . The GWAThe tested SNPs were analysed in a previous larger data set including other centres. For the current cohort, a screening and a verification test was performed in 1605 individuals. The comparison showed concordant results in 99% of analyses .The SNP analyses were performed with a MALDI-TOF mass spectrometer (SEQUENOM MassArray) using iPLEX reagents and protocol (SEQUENOM) and 10 ng DNA as PCR template. Primer sets were from Metabion . The laboratory methods have previously been described in detail .The genotypes for the SNPs were defined as: homozygous major allele (AA), heterozygote (Aa) and homozygous minor allele (aa). In cases with minor allele frequency (MAF) near 0.5, the same classification as that used in previous studies was used.Cases and controls were compared with regard to established and potential risk factors for breast cancer in order to identify possible confounders.An unconditional binary logistic regression model was fitted to analyse the association between SNPs and breast cancer. A per allele analysis was performed using a continuous variable with the values 0 (AA), 1 (Aa), and 2 (aa). The reported odds ratios (OR) with 95% confidence intervals (CI) denotes the risk difference when increasing the number or risk alleles with one. In addition ORs were calculated using the common allele (AA) as reference group for risk estimates for the separate genotypes in all analyses.Parity and age at first birth was dichotomised as nulliparous vs. parous and ≤ 25 years of age at first childbirth vs. age > 25 years of age at first childbirth. Overall breast cancer risk was calculated with the lowest groups as reference, adjusted for birth year and year at baseline.The SNP analyses were stratified on nulliparous vs. parous women and on age ≤ 25 years of age at first childbirth vs. age > 25 years of age at first childbirth. Parity and age at first childbirth were also studied in four categories.In addition, the material was stratified on single genotypes, and the breast cancer risk associated with parity and increasing age at first birth was calculated. These associations were reported using the p-values for the continuous analysis.All analyses were subsequently adjusted for matching criteria, age and year of inclusion in study, and for potential confounders. A potential confounder was defined as a factor with a distribution difference exceeding 5% units between cases and controls and only these factors were included in the multivariate analysis.In order to assess any potential interactions between selected SNPs and parity and between selected SNPs and age at first birth, an interaction-term was introduced in the logistic regression model. A p-value < 0.05 was considered statistically significant. In a third step, the p-value was corrected for multiple comparisons according to Bonferroni, i.e. divided by the number of comparison. In the present study, performing 28 interaction analyses, the corrected p-value regarded as statistically significant was 0.0018.As part of a sensitivity analysis, all analyses were repeated excluding women with information on less than 80% of SNPs in a single individual, as this may indicate a poor DNA quality. In these analyses 660 cases and 1310 controls were included.Among the described variables, cases were more often non-manual workers as compared to controls; 60% vs. 52% Table . More caA breast cancer association was seen for six of the 14 tested SNPs; rs2981582 (FGFR2) 1.28 (1.12-1.47), rs3803662 (TNRC9) 1.20 (1.04-1.39), rs12443621 (TNRC9) 1.19 (1.04-1.35), rs889312 (MAP3K1) 1.18 (1.02-1.36), rs3817198 (LSP1) 1.17 (1.10-1.35) and rs2107425 (H19) 0.86 (0.75-0.99) . For age at first birth, there was no breast cancer association (>25 years of age RR: 1.05 (0.87-1.28)).Only one interaction between the effects of the tested SNPs (rs981782 (5p12)) and parity was found with regard to breast cancer risk p = 0.02 Table . This asIn the per allele analyses, no clear patterns for risk associations were seen in the stratified analyses .The results of this present study are in agreement with previous GWAS studies for six out of 14 SNPs. With respect to breast cancer risk, there were no statistically significant gene-environment interactions between parity/age at first childbirth and SNPs and this is in agreement with the results in three other large-scale investigations -14.A total of 40% of the invited women in Malmö participated in the MDCS and women in the MDCS have been shown to have a higher incidence of breast cancer and they may also be selected towards a slightly higher socioeconomic status  than theParity and age at first birth were the main exposures of this study and were obtained from questionnaires answered at baseline. All women were 44 years or older at baseline, hence unlikely to have given birth to more children thereafter.SNP analysis method has been validated by repeating the analyses twice, in a subset and the reproducibility was very high . In ordeOverall, the sample size of this study was fairly small, yielding a statistical power issue. Many comparisons were made and there is a potential risk of a type I error. The replication of results concerning selected SNPs and breast cancer risk was based on previous studies, and all but one SNP showed associations in the expected direction . This strengthens the assumption that these results reflect true associations and were not only the result of multiple comparisons. As these analyses are made with an a priori hypothesis, the Bonferroni correction for multiple testing was not considered relevant. Concerning interaction analyses and the stratified analyses, these analyses were exploratory and hypothesis generating hence corrections for multiple comparisons for the performed interactions was considered valid. Due to few individuals in the analyses, the confidence intervals were wide and the statistical power was low which can have lead to a type II error. In order to address the risk of type II error, parity and age at first childbirth were dichotomized yielding larger study groups. Moreover, interaction term corrected for multiple comparisons (Bonferroni correction) yielded no statistically significant interactions.To our knowledge, four studies have been published studying breast cancer risk and the potential interaction between SNPs and parity/age at first childbirth -15. The The study by Travis et al. examined 120 gene-environmental interactions  categorising parity as nulliparous vs. parous and age at first childbirth as younger or older than 25 years of age in 7610 cases and 10 196 controls, making the results less vulnerable to type II error. They studied 12 SNPs and did not find any statistically significant interaction. Four of the SNPs examined in the present study (rs8051542 (TNRC9), rs12443621 (TNRC9), rs2107425 (H19) and rs7766585 (ESR1)) were not studied by Travis et al. [Milne et al. studied ten GWAS SNPs and two candidate SNPs associated with breast cancer in 26349 invasive breast cancer cases and 32208 controls with regard to interaction with reproductive factors. After adjustment for multiple comparisons no significant association was seen for parity  or age at first childbirth . Five of the studied SNPs by Milne et al. are the same as in this study: rs2981582 (FGFR2), rs889312 (MAP3K1), rs3817198 (LSP1), rs13281615 (8q24) and rs13387042 (2q35). The other seven SNPs studied by Milne et al. have been studied in more recent GWAS studies and are hence not included in the present study [Campa et al. studied 17 SNPs associated with breast cancer risk. When analysing gene-environmental risks including parity and other reproductive factors, no statistically significant association was seen. The study of Campa et al. included 8575 cases and 11892 controls, making this study less vulnerable to type II errors. Seven of the studied SNPs of Campa et al are the same as in the present study and one is on the same gene, making some of the results comparable to the present study.The results of this present study are in agreement with previous GWAS studies in SNPs and breast cancer risk for six out of 14 risk SNPs and is in agreement of null results for SNP parity/age at first childbirth interaction.None of the authors have declared any competing interests.SB carried out all the statistical analyses, participated in interpreting the results, reviewed the literature and drafted the manuscript. SH performed the SNP analyses, participated in interpreting the results and critically revised the manuscript. SiB participated in interpreting the results and critically revised the manuscript. MI performed the SNP analyses and critically revised the manuscript. GL participated in interpreting the results and critically revised the manuscript. JD participated in designing the study, critically revised the statistical methods and the manuscript. JC participated in designing the study and critically revised the results and the manuscript. JM designed the study, supervised and participated in interpreting all statistical analyses and critically revised the manuscript. All authors have read and given approval of the final manuscript.The study was carried out in the MDCS. In order to get access to the data, an application to the committee is needed. Please contact Ass. Prof. Jonas Manjer: jonas.manjer@med.lu.se.Appendix 1. Breast cancer risk in relation to selected SNPs stratified on parity. Appendix 2. Breast cancer risk in relation to selected SNPs stratified on age at first child-birth. Click here for file


Random periareolar fine needle aspiration (RPFNA) and ductal lavage (DL) are research techniques developed to (1) assess short-term breast cancer risk in asymptomatic women who are at increased risk for breast cancer and (2) track cytological response to risk reduction strategies. RPFNA and DL provide minimally invasive methods to repeatedly sample epithelial cells and research tools to investigate the biological origins of breast cancer in high-risk women. This review gives an overview of the strengths and limitations of both RPFNA and DL for risk assessment and breast cancer prevention in asymptomatic high-risk women. BRCA1/2 genetic mutation, or the presence of atypical hyperplasia. There is also evidence that cytological atypia in breast fluid or breast aspirate increases a woman’s risk of subsequently developing breast cancer. Random periareolar fine needle aspiration (RPFNA) and ductal lavage (DL) are minimally invasive research tools that are currently being utilized in a variety of clinical trials to test for the presence of cytological atypia in high-risk asymptomatic women and to track response to risk reduction strategies.A woman’s breast cancer risk increases with a first-degree family history of breast cancer, a Gail risk score of 1.66 or greater, Breast cancer incidence has been shown to be reduced in high-risk cohorts by chemoprevention agents such as tamoxifen and through prophylactic surgery –4. Howevsurrogate endpoint biomarkers [Biomarkers that vary with risk and response to prevention interventions are referred to as omarkers . As has omarkers , surrogaomarkers –10. CurrBreast tissue biomarkers offer the advantage of directly testing for precancerous changes in the breast. Atypia and lobular carcinoma in situ (LCIS) are associated with increased breast cancer risk . MoreoveRPFNA is a research technique that was developed by Carol Fabian, M.D., at the University of Kansas in the mid-1980s to (1) assess short-term breast cancer risk in women at high risk for breast cancer and (2) track cytological response to risk reduction strategies , 14. RPFIn 1986, the late Helene Smith proposed that breast cancer developed in a “high-risk field” or segment of the breast containing molecular changes that promote the development of a malignancy . The exiTM with 1 % formalin for 24 h. Carol Fabian’s group pools cells from both breasts into a single sample [The breast is anesthetized with 5 cc of 1 % lidocaine, 2–5 cm from the areola (depending on the size and shape of the breast), at approximately three and nine o’clock. Nine aspirations are performed per breast, with four aspirations from the medial skin site and five from the lateral breast site. While aspirations are guided by the tactile identification of stromal/ductal tissue, there is no attempt to aspirate a specific location within the breast ; 11–13, hyperplasia; 14–17, atypia; and >17, suspicious cytology [4 to 5 × 106 (Masood 11–13), (2) 106 to 5 × 108 (Masood 14–17), and (3) 107 to 109 (Masood >17). Accordingly, epithelial cytology from repeat samples possibly provides the ability to examine risk biomarkers and to prospectively predict the subsequent development of breast cancer.RPFNA epithelial cytology is most frequently evaluated on thin layer cytologic preparation using the Masood Cytology Index score, a semiquantitative, numeric assessment scale that is weighted toward epithelial nuclear morphology. A minimum of one epithelial cell cluster with >10 epithelial cells is required for assessment, and the most abnormal cell cluster is scored , 17. Epicytology , 17. TheIn 2000, Carol Fabian conducted a single institution study that demonstrated that cytological atypia in RPFNA independently predicted for an increase in breast cancer risk in high-risk women . In thisRPFNA has the advantages of being inexpensive, performed in a clinic exam room, and repeatable, allowing for the serial assessment of predictive biomarkers. The majority of RPFNA aspirates in high-risk premenopausal and perimenopausal women are cellular , 24. As p < .0001) and .7260 (p < .0001), respectively. Importantly, the reproducibility of duplicate RPFNA samples from the same breast was not affected by age, body mass index, 5-year Gail risk score, menopausal status, BRCA1/2 mutation, or the number of first-degree family members with breast and/or ovarian cancer [While RPFNA has gained in acceptance, its use is limited by the absence of data with respect to reproducibility in the technique and interpretation in multiinstitutional studies. To address these potential limitations, the CALGB Prevention Group tested the reproducibility of RPFNA in a multiinstitutional cross-sectional study of high-risk women •. In thin cancer •.Studies performed by the CALGB Prevention Group provide evidence that RPFNA measurements are reproducible in a cooperative group setting and support the future use of RPFNA in multiinstitutional prevention trials. However, there were several limitations to this study. First, the number of subjects was small (63 women). Second, intraoperator variability of RPFNA was assessed, but the interoperator variability was not. The concordance between RPFNA measurements provided evidence that single operators in multiple institutions can produce similar results. But this study did not address whether different operators would have similar findings in the same woman. Despite the limitations of this study, these data provide important validation of the reproducibility of RPFNA in a multiinstitutional cross-sectional study that included cohorts that varied in demographic composition •.2 DFMO or placebo p.o. once per day for 6 months, followed by a repeat RPFNA. Of 119 subjects entered, 96 % completed the study and were evaluable for the main study end point [Carol Fabian’s group pioneered the use of RPFNA-based biomarkers . The firnd point . The trind point . Since tnd point •, 23–30.nd point •, 32•. Up < .001), and therefore, offered a more sensitive method of detecting cellular atypia [DL is an experimental technology that utilizes nipple fluid to identify high-risk and occult malignant lesions in breast epithelial cells. Initial data from a multicenter study by Dooley et al.  establisr atypia . CytologThe procedure begins with breast massage and placement of a suction device to the nipple to identify fluid-yielding ducts and visualize non-fluid-yielding ducts, which are then cannulated with a microcatheter, infused with saline or physiologic solution, and then aspirated to collect cells from the lining of the milk duct Fig. . The locThe potential for early detection of breast cancer by DL was examined in 2004 by Khan et al.  in 39 woDL is a nonsurgical breast epithelial sampling procedure that was developed to enhance the cellularity and reproducibility for detection of epithelial atypia and for serial observation in chemoprevention trials. DL offers the ability to label individual ducts for future recannulation in order to decrease sample variance in prevention trials, in addition to identifying the location of any abnormal cells found by cytopathology. However, the efficacy of DL for the serial monitoring of breast epithelium was tested for reproducibility in a study of 23 women, indicating that fewer than 50 % of women producing atypical DL samples on the first attempt were found to have atypical samples on a repeat DL attempt . The relTolerability of DL has been reported as a possible limitation to its clinical use. The first multiinstitutional study of DL by Dooley et al.  reportedOptimal breast cancer predictive models are combining biologic and clinical markers to enhance the sensitivity of DL and identify characteristics of malignant changes within the breast. For example, DL was utilized for cancer detection in 2002 by King et al.  with 39 To determine the principle that biomarkers of chemoprevention agents can be examined using serial DL, several phase 2 prevention trials were initiated using tamoxifen. Bhandare et al.  assessedComparison of DL and RPFNA as tissue acquisition methods in early breast cancer prevention trials was tested prospectively at a single institution by evaluating sample adequacy and patient tolerability . In thisEvaluation of the cost effectiveness with RPFNA and DL in stratifying women for breast cancer preventive interventions was studied in 2004 and demonstrated effective means to motivate women at high risk for breast cancer to take tamoxifen if atypia were found. In this study, RPFNA was found to be both less expensive and more likely to produce an evaluable specimen, when compared with DL . FurtherCurrently, we lack agents to prevent estrogen receptor-negative (ER−) breast cancer. A wealth of targeted inhibitors are undergoing clinical testing for the treatment of ER(−) breast cancer. However, in order to effectively test targeted agents for prevention, we first need biomarkers to identify women who have the highest likelihood of response. Strategies to rapidly test for response and resistance are also needed.The past 20 years have focused on developing single biomarkers. However, it is clear that normal mammary gland homeostasis requires the coordinated regulation of phosphoprotein signaling networks (rather than single proteins). Currently, we lack an understanding of the biology of breast cancer initiation. The combination of RPFNA or DL with high-throughput proteomic profiling holds promise for investigating whether dysregulation of interconnected signaling networks predicts cancer initiation and progression. This approach may also allow for rapid tracking of response to complex risk reduction interventions.Low-toxicity targeted agents are potentially useful in the prevention of ER(−) breast cancer. However, emerging evidence suggests that some targeted agents, due to their specificity, may have unanticipated mechanisms of resistance . Recentl


It has been suggested that soy isoflavones have protective effects against breast cancer. However, data from epidemiological studies are not conclusive. A recent meta-analysis showed that soy intake was inversely associated with breast cancer risk in Asian but not Western populations, which indicates that protection against breast cancer may require that women consume levels of soy typical in Asian diets. In addition to the amount of soy isoflavones consumed, the form and food source of isoflavones, timing of isoflavone exposure, estrogen receptor status of tumors, and equol-producer status and hormonal profile of individuals may modify the association between soy isoflavone intake and the risk of breast cancer. These factors might explain the heterogeneity of results from studies. This present report contrasts background data from Japanese and Western women to identify the potential modifying of these factors. Isoflavones also inhibit the activity of key enzymes that convert androgens to estrogens.2 In addition, isoflavones have been shown to be anti-proliferative,3 proapoptic,4 anti-angiogenic,5 anti-oxidative,6 and anti-inflammatory.7 Taken together, these findings indicate that soy isoflavone may be a potent agent for preventing breast cancer. However, some data suggest that isoflavones promote breast cancer. Indeed, it has been shown that genistein stimulates the growth of estrogen-sensitive breast cancer cells in ovariectomized mice.8,9 Because of the apparent complexity of the relationships between isoflavones and breast cancer in laboratory studies, data from human studies are particularly important. However, current data from epidemiological studies are not conclusive. There is some epidemiological evidence of an inverse association between isoflavone and breast cancer, but it is inconsistent.10 Thus, no consensus has emerged regarding the preventive aspects of isoflavones.There has been much interest in the potentially protective role of soy in breast cancer development. Soybeans are the main source of isoflavones, which are classified as phytoestrogens. Laboratory data show that isoflavones have a wide range of biological actions. First, they have an affinity for estrogen receptors in vitro.11 A meta-analysis of 8 studies showed that Asian women consuming the highest amount (≥20 mg/day) of dietary isoflavones had a 29% reduction in breast cancer risk, as compared with those with low consumption (<5 mg/day) of isoflavones. An approximately 50% reduction in risk was reported in a prospective study of Japanese women.12 In contrast, a meta-analysis of 11 studies of women eating Western diets found no association between isoflavone intake and breast cancer risk. In that analysis, the median for the highest intake was 0.8 mg/day. Therefore, women may need to consume the amount of soy in Asian diets to gain protection against breast cancer. Alternatively, the protective effects may only be present when exposure to soy occurs early in life. Differences in (1) the form and source of isoflavones, (2) the biological response to isoflavones among ethnic groups, and (3) the interactions with individuals’ hormonal profiles may also explain the heterogeneity of the findings. This present report attempts to address these issues by contrasting background data from Japanese and Western women.Recently, Wu et al conducted 2 separate meta-analyses of studies carried out in Asian and Western populations.13,14 suggests that lifestyle changes, including a change in dietary isoflavone intake, may play a role. According to the Cancer Incidence in Five Continents Vol IX,15 the incidence rate for breast cancer in 1998–2002 was 32.0 per 100 000 among Japanese women in Osaka, Japan; 105.6 per 100 000 among white women in Hawaii; and 107.5 per 100 000 among Japanese women in Hawaii.The incidence rate of breast cancer is historically much lower in Japan than in Western countries. One possible explanation for this is that Japanese consume much more soy foods containing isoflavones. The fact that the incidence rate of breast cancer among Japanese immigrants rises as the length of time in the host country increases16 for more than 50 years. About 20 000 individuals from 6000 households in a randomly selected 300 districts are invited to participate in this survey every year. Foods and beverages consumed by a family are weighted and recorded for 3 consecutive days (1-day records have been used since 1995). According to the NNS reports, the mean intake of soy products was 67.2 g/day in 1975 and 57.7 g/day in 2005. Recently, the Food Safety Commission in Japan estimated isoflavone intake (expressed as aglycone) among Japanese, using the 2002 NNS data (response rate unknown). The median and the 95th percentile isoflavone intakes were 18 and 70 mg/day, respectively. Mean estimates of daily consumption reported in other studies in Japan ranged from 26 to 54 mg.17 A database of the isoflavone content of Japanese foods, based on a validated direct measurement method, was first reported by Kimira et al.18 In a report from the Takayama study—a prospective study of residents in Takayama city, Gifu, Japan19—the mean estimate, based on a validated food frequency questionnaire, was 19.6 mg per 1000 kcal among 15 724 female residents aged 35 years or older .20 Daily intake of isoflavones is estimated to be between 0.5 and 3 mg in the United States21,22 and less than 2 mg in European countries.23 In a recent study of 8809 adults from the National Health and Nutrition Examination Survey (NHANES), the mean isoflavone intake, based on one 24-hr diet recall, was 1.2 mg.24 Changing dietary habits have been reported in migrant populations. For example, soy intake among Japanese-Americans appears to be higher than that of US whites and lower than that of Japanese living in Japan. Takata et al25 reported that mean isoflavone intake, estimated from a food frequency questionnaire, was 11.6, 7.2, and 3.5 mg per 1000 kcal in Japanese women in Gifu, and Japanese and white women living in Hawaii, respectively. According to the 2002 NNS data in Japan, isoflavone intake was associated with age. Even among adults, younger age groups tended to consume less isoflavones: the reported mean daily intake of isoflavones was 19.4, 21.5, 23.5, 27.4, 33.6, and 29.6 mg per day for adults aged 20–29, 30–39, 40–49, 50–59, 60–69, and 70+ years, respectively. Soy isoflavone intake will likely decrease among Japanese in the future.The nutritional status of Japanese has been evaluated annually by the National Nutritional Survey (NNS)26 collected 24-hour urine and serum samples from Japanese women enrolled in the Japan Public Health Center-based (JPHC) prospective study. The mean levels for daidzein and genistein were 17.0  and 14.2 (14.1) µmol/day, respectively, in urine. The corresponding values in plasma were 119.9 (135.8) and 475.3 (510.4) nmol/L, respectively. These values were reasonably correlated with isoflavone intake estimated from food-frequency questionnaires or diet records (r = 0.22–0.48). Similar values in the urine, plasma, and serum of Japanese women have been reported in other studies.27–29 Even in spot urine samples, isoflavone levels were well correlated with intakes estimated from diet records (r = 0.30 for daidzein and 0.27 for genistein).30 In a study of 59 women aged 20 to 45 years, the median values were 4.0 and 3.2 nmol/mg creatinine for daidzein and for genistein, respectively, in first morning urine specimens.31 Another study of 419 women reported that the mean daidzein level was 12.8 nmol/mg creatinine in spot urine samples collected at approximately 2:00 PM.32 Among Western populations, reported urinary isoflavone levels have been much lower. The geometric mean levels were 0.27 and 0.08 nmol/mg creatinine for daidzein and genistein, respectively, in urine specimens of participants in the NHANES between 1999 and 2000.33 In a study of UK women, the geometric means for daidzein and genistein were 0.55 and 0.28 nmol/mg creatinine, respectively, in spot urine, and 7.9 and 15.2 µmol/L, respectively, in serum.34Biomarkers are also useful in estimating exposure to soy isoflavones. Yamamoto et al22 Such uses are increasing.35 Recently, new isoflavone datasets for non-soy based foods have been developed to estimate isoflavone intake more accurately. Horn-Ross et al22 assessed the isoflavone intake in 447 non-Asian women in the San Francisco Bay area, using a new database. Mean intakes of genistein and daidzein were estimated to be 1.5 and 1.3 mg/day, respectively. Major sources were tofu, doughnuts, soymilk, white bread, and canned tuna. Interestingly, doughnuts accounted for about 20% of the average of daily intake of genistein and 15% of daidzein intake. It is likely that Japanese women are also consuming isoflavones from such “hidden” sources of soy. Using several databases of non-soy foods, such as cereals, eggs, dairy foods, meats, fish, nuts, and vegetables,21,36–39 and assuming that processed meats and fish contain 2% soy protein by weight, isoflavone intake from these non-soy foods was estimated to be approximately 0.68 mg per 1000 kcal among the women in the Takayama study (unpublished data). This value is substantially lower than the isoflavone estimate from soy foods .Sources of isoflavones also differ between Japanese and Western populations. The intake of traditional soy-based foods is high in Japan. Based on data from the Takayama study, the most common soy foods are tofu, miso (soybean paste), soybeans, fried tofu, natto (fermented soybeans), and soymilk. The fermented soy foods (miso and natto) accounted for about 40% of total isoflavone intake. In Western populations, the consumption of traditional soy foods is substantially lower, and a portion of total isoflavone consumption is derived from soy protein and soy flour added to a variety of foods.39 Knowing the bioavailability of the different isoflavone forms is thus fundamental. Watanabe et al conducted a kinetic study of Japanese men after ingestion of 60 grams baked soybean powder (kinako).40 Differences in the form of isoflavone (aglycone vs glucoside) are believed to alter isoflavone pharmacokinetics and, hence, the association between soy intake and breast cancer risk. However, there is no consensus among studies on the bioavailability of aglycone vs glucoside.41 In addition to isoflavone form, the matrix in which it is delivered is likely to have an effect on pharmacokinetics. It is also possible that an individual’s background diet may modify the association of isoflavone intake with breast cancer risk by affecting the pharmacokinetics of metabolism of isoflavones. In addition, frequency of ingestion may affect bioavailability of isoflavones.41 An optimum steady-state serum isoflavone level would be expected from consuming relatively small doses of soy throughout the day, as opposed to a single dose at one time. Indeed, Gardner et al42 observed that plasma isoflavone concentrations remained constant when soy foods were consumed at breakfast, lunch, and dinner.In fermented soy products, like natto or tempeh, aglycones are the principal form of isoflavone, whereas in unfermented soy products, like soy milk or soy supplements, the glucoside form is predominant.43 Equol, a metabolite of daidzein, has been identified in the urine and blood samples of some but not all humans. Intestinal bacteria play an essential role in daidzein metabolism. Several candidate bacteria for daidzein metabolism, including one reported by Ueno et al,44 have been suggested, but the human intestinal bacteria responsible for daidzein metabolism are yet to be identified.45 Equol binds with greater affinity to estrogen receptors than its parent compound daidzein,46 and it is superior to all other isoflavones in its antioxidant activity,47 which suggests that its overall potency is greater. The prevalence of equol producers appears to be higher in Japanese than in whites,27,48,49 which might explain the additional benefits conferred on Japanese women in terms of reduced breast cancer risk.Metabolism of isoflavone can vary greatly between individuals, even when they consume an identical amount of isoflavone-containing food.50 possibly because of alterations in gene expression and morphological changes in the mammary glands. The details have been described elsewhere.51 To this author’s knowledge, 3 case-control studies have examined the association between soy intake during childhood or adolescence and the risk of breast cancer among Asian women.52–54 All reported a significant inverse association. Another case-control study of Canadian women also observed a significant inverse association between adolescent isoflavone intake and breast cancer risk.55 A recent prospective study of Chinese women confirmed that a high intake of soy foods during adolescence was associated with a reduced risk of breast cancer.56 Unfortunately, there has been no such study among Japanese women.In vivo studies have consistently shown that prepubertal or pubertal exposure to genistein reduces the incidence and/or multiplicity of chemically induced mammary tumors in experimental animals,57 Some epidemiological studies observed associations between pre- and perinatal characteristics, such as birth order and birth weight, and subsequent risk of breast cancer.58 There has been no study assessing the association between maternal soy intake during pregnancy and the risk of breast cancer in offspring; however, by altering the estrogen environment, soy exposure in utero may affect the subsequent risk of breast cancer.Perinatal factors are also thought to influence subsequent breast cancer development. Trichopoulous hypothesized that the developing breast is influenced by the fetal environment, particularly variations in hormone concentrations, which could mediate subsequent breast cancer development.59 reported that in 7 Japanese women at delivery the mean levels of genistein and daidzein were 83.9 and 45.5 nmol/L, respectively, in maternal plasma, and 165 and 58.8 nmol/L, respectively, in umbilical cord plasma. They observed high correlations between levels in maternal and umbilical plasma . A study by the present author, which included 194 pregnant women, supported the hypothesis that isoflavone can be transferred from mother to fetus.30 In that study, we also included an assessment of maternal dietary intake of soy during pregnancy and measurements of estrogen levels in maternal and umbilical cord blood; the mean soy isoflavone intake was 21.7 mg per day. The geometric mean levels for serum genistein and daidzein were 116.5 and 50.2 nmol/L, respectively, in maternal blood at delivery, and 126.9 and 38.6 nmol/L, respectively, in umbilical cord blood. Estradiol level in umbilical cord blood was unrelated to isoflavone levels in both maternal and umbilical cord blood.In utero soy exposure occurs among Japanese. Adlercreutz et al60 reported that soy-based formulas contained approximately 32 to 47 µg/ml of isoflavone, indicating that a 4-month old infant fed soy formula would consume 28 to 47 mg of isoflavones per day. In their study, the mean plasma concentrations of genistein and daidzein were 2530 and 1160 nmol/L, respectively, in 7 infants fed soy-based formulas. An isoflavone intake of 3 mg for Japanese infants is not low, if we take into account the body size of babies; however, it would be much lower than that available from a soy-based formula. Badger et al61 reported substantial differences between exposure to soy between Japanese and Americans. If these differences are real, then Japanese fetuses would be exposed to the same high isoflavone levels that were present in maternal circulation. Serum isoflavone levels among Japanese infants would then be expected to decline at birth and remain low, until soy foods were introduced as baby food. Thereafter, the level would increase to the usual adult levels as a result of habitual intake. American infants fed soy formula would be expected to have high serum levels starting immediately after birth and persisting until they were weaned, at which time levels would drop and likely remain low thereafter.Infants in Japan are weaned on to soy products between 6 and 12 months of age, after which they continue to receive isoflavone-containing foods indefinitely; tofu and miso soup are common baby foods. Based on intake frequencies of tofu, miso soup, natto, soybean flour, and soymilk reported by 288 mothers, 6-month-old Japanese infants consumed about 3.1 mg of soy isoflavone per day (unpublished data). Setchell et al62,63 Because isoflavones have an affinity for estrogen receptors, the protective effect of soy intake may be pronounced for estrogen receptor-positive tumors. To date, 6 studies have assessed the association of soy or isoflavone with breast cancer risk, with regard to the estrogen receptor status of tumors.34,64–68 Some of these studies, including 1 conducted in Japan, observed an inverse association of soy intake with the risk of estrogen-positive tumors,64,65,67 but not with the risk of other tumor types.34,66,68 The incidence rate of estrogen receptor-positive tumors is increasing in the United States.69 In addition, it has been reported that the prevalence of estrogen receptor-positive expression in normal breast tissue from Japanese women was lower than that in white women.70 This may simply be a result of the lower incidence of breast cancer among Japanese women. The determinants of estrogen receptor status in normal breast tissue and tumors are not known.Epidemiological studies have found that several hormone-related lifestyle factors, such as nulliparity, earlier age at menarche, higher body mass index (BMI), and use of postmenopausal hormones were related to elevated risk of estrogen receptor-positive, but not estrogen receptor-negative, cancers.71 Evidence from case-control studies has supported a protective role for soy in premenopausal women versus postmenopausal women.10 The effects of soy and isoflavone on the level of circulating estrogen have been examined in many intervention studies. A recent meta-analysis of intervention studies revealed that although soy and isoflavone consumption did not affect estradiol or estrone, it did reduce FSH and LH in premenopausal women.72 Moreover, in postmenopausal women, soy and isoflavone were associated with a small but nonsignificant increase in total estradiol. These observations lend tentative support to the above hypothesis. However, in a cohort study that observed a significant inverse association between soy isoflavone intake and breast cancer risk among Japanese women, the association was somewhat stronger in postmenopausal women than in premenopausal women.12 A similar tendency was noted in another recent cohort study of Asian women.68 In addition, some studies have observed lower endogenous estradiol levels in low-risk populations than in high-risk populations.73,74 Although postmenopausal estrogen levels should not be compared when measured by different laboratories, estrogen levels appear to be low among postmenopausal Japanese women .75–79 Such a low-estrogen environment, which likely reflects low body fat mass, may favor the protective effects of soy on breast cancer in Japanese women. Some studies of soy and breast cancer risk included analyses stratified by BMI but not by estrogen level. The results were not consistent: one observed a somewhat stronger association in women with high BMI,55 but others noted no significant differences.68,80It has been suggested that isoflavones have stimulatory effects in low-estrogen environments, and that in high-estrogen environments, they block the effects of estrogen.The causes of breast cancer remain unclear. Established risk factors include age, race/ethnicity, reproductive factors, and obesity. Unfortunately, there are few modifiable factors. Dietary factors have been implicated in the etiology of breast cancer and soy/isoflavone has been a candidate for dietary intervention. However, existing evidence from epidemiological studies is not conclusive and there have been few prospective studies of the issue. Estimation of soy and isoflavone intake, like other dietary components, is subject to measurement error. In addition, the existence of hidden sources of soy makes it difficult to estimate total isoflavone intake accurately, especially among “nonconsumers” of soy-based foods.There is a need for more, methodologically sound, prospective studies—with extensive exposure measurement. To interpret the data, we need to consider the factors described above, namely, isoflavone dose, forms and sources of isoflavone, timing of isoflavone exposure, and the equol-producer status, estrogen-receptor status, and hormonal profile of individuals, as these are all factors that potentially modulate the association between soy intake and breast cancer risk. Other dietary, environmental, and genetic factors may also modify the association. Future studies need to address these questions by including samples large enough to detect the factors that are capable of modifying the associations between soy and breast cancer risk.


Breast cancer is a complex disease and may be sub-divided into hormone-responsive (estrogen receptor (ER) positive) and non-hormone-responsive subtypes (ER-negative). Some evidence suggests that heterogeneity exists in the associations between coffee consumption and breast cancer risk, according to different estrogen receptor subtypes. We assessed the association between coffee consumption and postmenopausal breast cancer risk in a large population-based study , overall, and stratified by ER tumour subtypes.Odds ratios (OR) and corresponding 95% confidence intervals (CI) were estimated using the multivariate logistic regression models fitted to examine breast cancer risk in a stratified case-control analysis. Heterogeneity among ER subtypes was evaluated in a case-only analysis, by fitting binary logistic regression models, treating ER status as a dependent variable, with coffee consumption included as a covariate.> 5 cups/day compared to OR≤ 1 cup/day: 0.80, 95% CI: 0.64, 0.99, P trend = 0.028). In the stratified case-control analyses, a significant reduction in the risk of ER-negative breast cancer was observed in heavy coffee drinkers  in a multivariate-adjusted model. The breast cancer risk reduction associated with higher coffee consumption was significantly higher for ER-negative compared to ER-positive tumours (P heterogeneity (age-adjusted) = 0.004).In the Swedish study, coffee consumption was associated with a modest decrease in overall breast cancer risk in the age-adjusted model (ORA high daily intake of coffee was found to be associated with a statistically significant decrease in ER-negative breast cancer among postmenopausal women. Coffee is one of the most popular beverages in the world. The latest coffee trade statistics estimated that world coffee production amounted to 7.4 billion kg in 2009/2010 [Coffee is interesting in the light of breast cancer etiology because of its complex make-up of chemicals, several of which have been shown in experimental studies to have cancer risk altering potential through meaningful biological mechanisms. The scientific community, however, stands divided over toxicity of the beverage. It has been demonstrated in experimental and clinical studies that coffee, being a complex mixture of caffeine and polyphenols -7, can pBreast cancer is a complex disease and may be sub-divided into hormone-responsive (estrogen receptor (ER) positive) and non-hormone-responsive subtypes (ER-negative). Coffee itself might contain compounds that differentially affect breast cancer of different ER subtypes. We, thus, hypothesize that heterogeneity exists between coffee consumption and breast cancer risks for ER-positive and ER-negative breast cancers.In this study, we examine the association between coffee consumption and postmenopausal breast cancer risk in a large population-based study , overall and stratified by hormone receptor status. As the possibility that the weak relationship between high levels of coffee consumption and the occurrence of breast cancer is due to confounding by related dietary or lifestyle factors is tenable , we alsoin situ of the cervix) . The final study group consisted of 2,818 cases and 3,111 controls. The ethical review board at the Karolinska Institute and the six ethical review boards in other regions of Sweden approved the study.Subjects were drawn from a population-based case-control study, which has been described in detail previously . The paret al. [in situ tumours  10: C50 and D05) and 6,657 controls, frequency matched by year of birth and study region. Two controls per case were randomly selected from the lists of residents provided by the population registries. For the present analysis, in situ cases were excluded. The study was approved by the ethics committees of the University of Heidelberg and the University of Hamburg. All study participants gave written informed consent.For the validation analysis, subjects were drawn from the population-based case-control MARIE (Mamma Carcinoma Risk factor Investigation) study which was carried out from August 2002 to September 2005 in two study regions in Germany (the Hamburg and Rhein-Neckar-Karlsruhe regions). Details of the study design can be found in Flesch-Janys et al. . Brieflyth percentile of the age of natural menopause , or otherwise as unknown. Subjects classified as post-menopausal in this manner (280 cases and 303 controls) were assigned an age at menopause according to their case/control and current smoking status corresponding to the mean age at natural menopause in the respective groups.Data were obtained by means of an extensive mailed questionnaire requesting detailed information on established and possible breast cancer risk factors, including reproductive and menstrual history, family history of breast cancer, hormone replacement therapy (HRT) and anthropometric measures, such as body mass index (BMI). Information on lifestyle such as smoking (> 1 year or > 100 cigarettes), alcohol intake (g/day) and physical activity  was also collected from the questionnaire. Highest education level attained was available as a categorical variable . Data on the consumption of coffee one year prior to interview, specified in cups per week, where a cup was equivalent to 1.5 dl, were also collected. Age at menopause was defined as the age of the last menstrual period or age at bilateral oophorectomy, if one year or more prior to data collection. The women were considered pre-menopausal if menopause occurred less than one year before data collection. Women with hysterectomy, menses due to HRT or missing information were considered post-menopausal if they had reached the 90Information regarding the retrieval of hormone receptor status from the medical records of all participants from surgical and oncological units throughout Sweden has been presented in detail elsewhere ,21. AlthFor the validation study (MARIE), information on potential risk factors for breast cancer was obtained in face-to-face interviews using a standardized questionnaire. Nutritional data were collected using a food frequency questionnaire with 176 food items regarding dietary habits in the year prior to date of diagnosis for cases and date of food frequency questionnaire completion for controls. The consumption of caffeine-containing coffee was calculated in cups per day based on the information on both portion size  and frequency . The analysis was limited to women who answered both questions on portion size and frequency of caffeine containing coffee consumption. The final study group comprised 5,395 controls and 2,651 cases. Information on tumour characteristics, such as ER and PR status, was obtained from medical records.The variable for coffee consumption was categorized as follows: one cup or less per day; more than one to three cups/day; more than three to five cups/day; five or more cups/day. These categories were based on the distribution within the control group. Since very few women abstained from coffee, we combined abstainers and low consumers (one cup per day) into a single category. Women who consumed one cup or less of coffee per day served as the reference group for all regression analyses.Unconditional logistic regression models, adjusting for the matching factor, age at enrolment in years (continuous), were applied to evaluate if established or possible breast cancer risk factors had (including coffee consumption) significantly different distributions/means  between breast cancer cases and controls in this study.2) and alcohol consumption (g/day)), logistic regression analysis, for binary risk factor variables  or proportional odds logistic regression, for categorical risk factor variables , highest education level , and recent physical activity ). The Wald test was used to determine the statistical significance of an overall linear trend for the association between coffee consumption, treated as a semi-continuous variable, and the breast cancer risk factor in the models fitted.The relationships between coffee consumption and other breast cancer risk factors were explored in the control population by treating coffee consumption as a covariate and using linear regression analysis for continuous risk factor variables (age at menarche (years), age at menopause (years), BMI , adjusted for age at enrolment, HRT, smoking and education, and adjusted for age at enrolment, HRT, smoking, education and daily alcohol consumption. The Wald test was used to determine the statistical significance of an overall linear trend for the association between coffee consumption, treated as a semi-continuous variable, and the breast cancer risk.P-value for association < 2.0 × 10-16), we assessed the extent to which coffee consumption drives each of the two tumour characteristics, by fitting multinomial regression models for five outcomes . We compared a model without parameter restrictions to models with parameters restricted such that coffee consumption was only allowed to be associated with one tumour characteristic at a time. Likelihood ratio tests, with two degrees of freedom, were used to test the null hypothesis that associations between coffee consumption and PR status was due only to an association with ER or PR status.Since ER and PR status are strongly correlated , treating ER or PR status as dependent variables, with coffee consumption included as a covariate. ORs and corresponding 95% CI were estimated for each coffee consumption category. iagnosis , and thaiagnosis , every mThe validation analysis based on the MARIE study population was performed using Proc LOGISTIC in SAS version 9.2 . The variable on coffee consumption was categorized in the same way as in the Swedish study with women who consumed one cup or less of coffee per day as the reference group. Unconditional logistic regression models were used to estimate ORs and corresponding 95% confidence intervals. To test for trend, we treated the four categories of cups per day as a continuous scored variable in the model statement only.P-values presented are two-sided tests of statistical significance at the 5% level.All statistical computations for the Swedish study were performed using R version 2.8 . All P-vP = 0.057 in Swedish samples and P = 0.0026 in MARIE samples), a result consistent with the literature [P = 0.176).Table terature . Family P = 0.008), smoking (P < 0.0001) and highest education level attained (0.041).Table > 5 cups/day: ≤ 1 cup/day: 0.80 , P = 0.028). When the model was further adjusted for HRT, smoking, education and average daily alcohol consumption, the protective effect on overall breast cancer risk was no longer found to be statistically significant. In the stratified analyses, significant reductions in the risk of ER-negative and PR-negative subtypes were observed, with the strongest effect being seen in ER-negative subtypes, for all models examined , P = 0.0003).Table P heterogeneity (age-adjusted) = 0.004). The effect of coffee consumption was, however, not significantly different by PR status (P heterogeneity (age-adjusted) = 0.230). The fitted multinomial logistic regression model  without parameter restrictions was not found to perform better than the model with coffee consumption effect, restricted to be independent of PR status (P = 0.877). On the other hand, the unrestricted model performed better than the model with coffee consumption effect restricted to be independent of ER status (P = 0.034).We next tested for heterogeneity in the effects of coffee consumption on hormone receptor status (ER and PR) in the Swedish study. The breast cancer risk reduction associated with higher coffee consumption was significantly higher for ER-negative compared to ER-positive tumours , P trend = 0.173). Although the difference in overall effect sizes for different breast cancer subtypes were less impressive in the validation dataset, the strongest protective effect was similarly observed for the ER-negative subtype , P trend = 0.326), followed by the PR-negative subtype , P trend = 0.280). The ORs, corresponding 95% CI, and P-values for trend were not altered by the introduction of smoking, alcohol consumption and other lifestyle factors.Motivated by the trend test results in the ER-negative cancers, we performed a validation analysis using the MARIE study. Table Our main finding was that coffee consumption was associated with a strong reduction in breast cancer risk for the ER-negative tumour subtype. This effect was independent of HRT, smoking, highest education level attained, and average daily alcohol consumption.P = 0.0003) and 33% (P = 0.034) less likely to get the ER-negative and PR-negative disease, respectively, than the reference group. The effects were also found to be independent of PR status. Motivated by the trend test results of the ER-negative breast cancer subgroup using the Swedish data, we attempted to validate the results in the independent MARIE study, conducted in Germany. Though not reaching statistical significance, the strongest protective effect from heavy coffee consumption was similarly observed for the ER-negative subtype , P = 0.326) in the validation study.In the multivariate-adjusted Swedish study, women who drank more than five cups of coffee per day were 57%  and highest education level attained (P = 0.0003). The coffee consumption variable and risk factor data of both studies were self-reported, and could thus be subjected to errors in measurement. However, correlations between data collection via food frequency questionnaires and weekly diet records are generally high [A limitation of our study is that receptor status was available for only 65.4% of the Swedish population. We have compared the characteristics between breast cancer cases with and without ER status (Table S1 in Additional file lly high ,32. Cofflly high , decaffeStrengths of our study include it being a population-based study, its large sample size and detailed information on relevant variables: coffee consumption, reproductive and hormonal risk factors, and tumour characteristics. We have also obtained supporting evidence in a large, well-described and independent population-based study.In conclusion, we found no evidence that coffee consumption increases the overall risk of postmenopausal breast cancer. However, a high daily intake of coffee was found to be associated with a significant decrease in ER-negative breast cancer among postmenopausal women. Future studies are needed to confirm the effects of coffee consumption in the light of breast cancer subtypes.BMI: body mass index; BRCA1: breast cancer 1, early onset; BRCA2: breast cancer 2, early onset; CI: confidence interval; ER: estrogen receptor; HRT: hormone replacement therapy; ICD: International Classification of Diseases; MARIE: Mamma Carcinoma Risk factor Investigation; OR: odds ratio; PR: progesterone receptor.The authors declare that they have no competing interests.JLi, KH, KC, JLiu and PH designed the study. PS, DFJ and JCC headed the validation effort. JLi, PS and KH conducted the statistical analysis. JLi drafted the manuscript, with substantial contributions from all authors mentioned.Table S1. Descriptive characteristics of post-menopausal women with information on hormone receptor status and without.Click here for file


A large proportion of breast cancer cases are thought related to environmental factors. Identification of specific geographical areas with high risk (clusters) may give clues to potential environmental risk factors. The aim of this study was to investigate whether clusters of breast cancer existed in space and time in Denmark, using 33 years of residential histories.Q-statistics were used to identify space-time clusters of breast cancer. All analyses were carried out with both control groups, and for 66% of the study population we also conducted analyses adjusted for individual reproductive factors and area-level socioeconomic indicators.We conducted a population-based case–control study of 3138 female cases from the Danish Cancer Registry, diagnosed with breast cancer in 2003 and two independent control groups of 3138 women each, randomly selected from the Civil Registration System. Residential addresses of cases and controls from 1971 to 2003 were collected from the Civil Registration System and geo-coded. In the crude analyses a cluster in the northern suburbs of Copenhagen was consistently found throughout the study period (1971–2003) with both control groups. When analyses were adjusted for individual reproductive factors and area-level socioeconomic indicators, the cluster area became smaller and less evident.The breast cancer cluster area that persisted after adjustment might be explained by factors that were not accounted for such as alcohol consumption and use of hormone replacement therapy. However, we cannot exclude environmental pollutants as a contributing cause, but no pollutants specific to this area seem obvious. With more than one million new cases each year, breast cancer is the most common cancer among women, accounting for one-fifth of all new female cancer cases worldwide. The indThe majority of the established breast cancer risk factors are related to oestrogens. Early menarche and late menopause increase the risk, while reproductive factors such as many child births and young age at first birth reduce the risk. Hormone replacement therapy (HRT) for menopause, ionising radiation, alcohol intake, night shift work and some specific genetic mutations are also established risk factors-9. FurthMigrant studies of breast cancer found that women, who migrate from areas of low risk to areas of high risk, adopt the higher risk in the host country within a few generations,12, and The Danish Data Protection Agency (2007-41-0437) approved the study. In accordance with Danish law written consent was not obtained as the study was entirely register-based and did not involve biological samples from, or contact with study participants.th Revision of the International Classification of Diseases. Only primary cancers were included, however previous diagnoses of non-melanoma skin cancer were allowed. The study included 3138 cases in total.Female breast cancer cases were identified in the virtually complete population-based Danish Cancer Registry, to which it has been mandatory to report all new cancer diagnoses since 1987. We inclFemale controls were randomly selected from the Danish Civil Registration System using inWe used the unique personal identification numbers of cases and controls to trace residential histories from 1971 to date of diagnosis of cases and index date of their matched controls by record linkage with the Danish Civil Registration System. Recording of residential data in the civil registration system was not complete before 1971; hence this year was used as the cut-off point. We identified 45,916 unique addresses, each with a unique identification number composed of a municipality code, a road code, and a house number. The dates of moving in and leaving each residence were registered. The addresses were then linked to a register of all official addresses in Denmark, resulting in geographical coordinates for 45,404 of the residential addresses, and missing coordinates for the last 512 (1%) addresses. In the geocoding procedure, 86% of the addresses of both cases and controls matched to the exact house. Four percent matched to one of the neighbouring houses, 2% matched to the centre of the road, and 7% matched at the municipality level, which means that centroid coordinates of the municipality were assigned to these addresses. Equal proportions of addresses of cases and controls were geocoded in each of these categories. The ages of cases and their matched controls were calculated at the beginning and end of each residence, which enabled us to use different time scales in the spatio-temporal cluster analyses.We obtained reproductive data for all cases and controls born in 1935 and onwards using record linkage of the personal identification numbers of cases and controls to the Medical Birth Register and the From Statistics Denmark we obtained information on socioeconomic indicators aggregated in a 100 meter x 100 meter grid cell net covering all addresses in Denmark. Cells contained average values on income and education in 2008 for a minimum of 100 households at the time. The income variable was based on the yearly disposable household income, while the educational level was based on the person with the highest education in each household. These area-level aggregated measures of income and education were linked with cases and controls based on their most recent residential address.Q-statistics in the software called SpaceStat  to investigate potential space-time clusters of breast cancer. The method has been extensively described in previous studies[We used  studies,26,27. Bi and j, ci and cj are defined to be 1 if and only if a case, and 0 otherwise. The termj is a k nearest neighbour at time t of participant i; otherwise it is 0.k based on the fact that an individual can have up to k unique nearest neighbours. This statistic is recalculated for each case every time there is a change in place of residence. In the present study we also calculatedk is specified by the user; however there is no standard method to determine the optimal value of k.Where for individuals Q-statistics was determined by randomly assigning the case–control status to the residential histories under the null hypothesis of no association between places of residence and case–control status. Monte Carlo simulations were used to generate the distributions for hypothesis testing, and the randomization procedure was repeated over 999 iterations, resulting in a minimum p-value of 0.001.The statistical significance of the k), age at first child birth (continuous), number of children (continuous), area income (continuous) and area education level (continuous) in the model. The risk estimates were then converted into probabilities that a location would be assigned case status as a function of the reproductive and socioeconomic factors. These probabilities were used for adjustment in the spatio-temporal analyses. ConsequQ-statistics. We used a Bernoulli model in SaTScan, and the p-value for test of significance was obtained from Monte Carlo simulations (999 replications). We analysed elliptical clusters with a maximum cluster size of 15% of the total population and with the option “No Cluster Centers in Other Clusters”.We ran both unadjusted and adjusted analyses and all analyses were conducted twice, first with control group 1 and then with control group 2. Calendar year and age were applied as two different underlying time scales. Finally, we analyzed data with the two control groups combined in a 1:2 individually matched design. We repeated selected analyses of potential clusters in SaTScan (version 9.1.1),29. ThesFor clusters that were consistently found across both control groups and in several analyses, we calculated the relative risk of breast cancer associated with a residential history (minimum 5 years) inside the cluster area. Further, we examined if age and extent of tumour at date of diagnosis were different for cases living inside versus outside the cluster areas.The study included 3138 cases of breast cancer and two independent control groups with 3138 controls in each. The average age at diagnosis for cases was 63 years, and both cases and controls lived at 4.8 addresses, on average, during the period 1971–2003. Reproductive and socioeconomic data were available for 2078 of the cases and 4155 of the controls corresponding to 66% of the study population; descriptive statistics are summarized in Table k, size of study population and adjustment for covariates.Figure Q-statistics) using k = 25  as detected when calendar year was applied.With age as the underlying time scale, application of each of the control groups identified clusters in the area north of Copenhagen at several levels of Figure k and there was better agreement on the location of the area across the two control groups (results not shown). A small cluster of 1–3 statistically significant cases was detected in Odense with the second and the combined control groups, but it was too small to be regarded a true cluster (not shown).Figure When analyses were adjusted for reproductive and socioeconomic factors, only the combined control group identified a cluster north of Copenhagen Figure b. The coFinally, 138 cases and 203 controls had resided inside the northern Copenhagen cluster area for at least five years, resulting in a relative risk of breast cancer of 1.39 (95% CI: 1.11-1.74) for women who had lived within the area compared to those who had not. Cases from the cluster area were on average seven years younger and had fever metastases at time of diagnosis than cases living outside the Copenhagen cluster area.This population-based case–control study consequently found a statistically significant cluster of breast cancer in an area comprising the northern suburbs of Copenhagen present at almost the entire study period. A second cluster was found in Odense; however, this cluster was less evident.k and confirmed by supplementary analyses in SaTScan. The cluster area persisted in crude analyses restricted to the 66% of the study population with data on reproduction and socioeconomic indicators. However, when analyses were adjusted for reproductive and area-level socioeconomic factors the cluster area was smaller and only identified with the combined control group. This could suggest that the clustering of cases in this area is caused by geographical differences in reproductive and/or socioeconomic factors. But as the cluster area did not disappear entirely as a result of the adjustment it is also possible that other factors have contributed to the cluster. Further, with the cluster being persistent when the control groups were combined but not when they were used separately could also indicate that sample size has influenced the results.Clusters in the northern suburbs of Copenhagen were consistently identified spatially and temporally by use of each of the two control groups and when control groups were combined into one. Further, the cluster area was found at several levels of Results also suggested a small cluster of breast cancer cases in Odense; however, this area was only statistically significant when the second and the combined control groups were applied, and when the study population was reduced to 66%, the area only had 1–3 statistically significant cases, which, according to a previous simulation study, is too Q-statistics. Results of selected analyses in SaTScan with the second and the combined control group agreed on this area, however since the area was not found by Q-statistics with the first or the combined control group nor by SaTScan using the first control group, we regard this a chance finding.Finally, a cluster was also found in the Høje Taastrup area (south-west of Copenhagen), with the second control group and In the space-time cluster analyses we modelled time both as calendar year and as age, because if age-specific susceptibility exists in the development of breast cancer it might be revealed by use of the age time scale. In geneBreast cancer is one of the few cancers that is associated with high socioeconomic status. HoweverPrevious studies have found that high socioeconomic status at the individual- and at the area-level independent of each other were associated with higher risk of breast cancer,36. The Finally, we cannot exclude that other factors  with geographical variation might have contributed to the observed cluster. Since the cluster persisted through time, the responsible factor(s) are likely to have similarly persisted across many years. The area north of Copenhagen, where the cluster was detected, is mainly a residential area with single family houses, green spaces, forests and lakes. The area has lower population density than Copenhagen City, but higher than in municipalities further away from Copenhagen. Farming and heavy industry is not present in the area; but a highway and some major roads, railways and power lines intersect the area. However, large parts of Denmark have such infrastructure, therefore it seems unlikely that these factors could be related to the clustering of cases.Several previous studies have identified clusters of breast cancer, however it is difficult to compare these to the present study because they were conducted in other populations (in the US and Canada) and they use different methods and types of data. One previous study found high breast cancer mortality in a large region covering New York and Philadelphia metropolitan areas, using the scan statistic of SaTScan. ApplyinA recent Canadian study applied the scan statistics of SaTScan and found excess incidence rates of breast cancer in five counties in southern Ontario, which were suggested related to environmental pollution from industry and farming characterising these areas. The stuThe present study is among the first examinations of clusters of breast cancer in both space and time using residential histories. Cases were identified in the virtually complete high-quality population-based Danish Cancer Registry,39, thusk in the analyses. But SaTScan spatial scan statistics, which search for clusters using variable-sized scanning windows, were used to confirm the clusters suggested by the Q-statistics. The Q-statistics employ local space-time statistics for each individual and the question of inference with multiple tests arises. To address multiple testing we used two approaches driven by our simulation study[Q-statistics, such that a possible true cluster warrants further investigation if four or more significant cases were detected in the same area with a Qip = 0.001 and Qitp ≤ 0.05. The second was to use SaTScan to corroborate clustering in the time periods found significant under the Q-statistics. While not strictly a multiple testing correction, using SaTScan to corroborate results increased our confidence in the finding provided by the Q-statistics, further reducing the possibility of a false finding due to multiple testing.The lack of data on alcohol intake and use of HRT limits our ability to interpret the likely causes of the cluster in our study, as these known risk factors may explain at least part of the detected cluster; however the inability to adjust for all known risk factors is a shortcoming of almost all cluster studies. Due to no more than 33 years of residential history data we did not have the possibility to detect clusters that could have occurred early in life or young adulthood, which is an important limitation of the study. Furthermore, seven percent of the residential addresses were geocoded at the municipality level, which could have introduced some uncertainty to the study; however sensitivity analyses that omitted these less precise addresses indicated that results were not influenced by the geocoding uncertainty. Another limiting factor was related to computational time. Due to the large data sets of residential histories, a single analysis took up to 8 hours, thus we could not explore a very wide range of different levels of on study. The firThe present study found a space-time cluster of breast cancer in the municipalities north of Copenhagen based on 33 years of residential histories. The cluster was consistently found across two independent control groups, but after adjustment for reproductive factors and area-level income and education the cluster became less evident. The remaining less evident cluster might be explained by socioeconomic factors that were not accounted for such as individual-level income and education, alcohol consumption and HRT use. We cannot exclude environmental pollutants as a contributing cause, but no pollutants specific to this area seem obvious.HRT: Hormone replacement therapy; CG: Control group.Geoffrey M. Jacquez has an interest in BioMedware, the developer of the SpaceStat software used in this study. This has not influenced interpretation of the results and does not alter the authors’ adherence to all of the policies of BMC Cancer. The authors declare that they have no further competing interests.All authors have contributed to study design and interpretation of results. RBN preformed data collection, carried out the analyses, and wrote the paper. GMJ and JRM developed the method and provided analysis tools. AHP, JRM, AKE, GMJ and ORN critically revised the manuscript. All authors have read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/14/255/prepub


Breast cancer is the most frequently diagnosed cancer globally, and studies provide contradictory results about the possible effects of vitamin supplementation to reduce cancer risk. Our aim was to conduct a review to better investigate whether vitamin supplements given orally modify breast cancer risk.We conducted a comprehensive, systematic bibliographic search of the medical literature to identify relevant studies. Case-control, cohort studies, and randomised controlled trials (RCTs) published up to August 2013 that reported cancer risk estimates for vitamin supplementation were included. For each study, we retrieved study characteristics, study population, exposure evaluation, and risk estimates.We identified 26 studies  and overall, we found 104 estimates. We grouped all the estimates into six supplementation categories: vitamin A and beta-carotene, B-group vitamins and folic acid, vitamin C, vitamin D, vitamin E, and multivitamins. Only a few studies showed a statistically significant association between the consumption of supplemental vitamins and the occurrence of breast cancer, and most of the significant estimates were found in case-control studies. The results found in prospective studies seem to be in the opposite direction.The role of vitamin supplements in preventing breast cancer still remains unclear, considering our review. Although biologic mechanisms exist to support the anticancer effects of vitamins, there is no clear evidence for an effect in cancer prevention for vitamin supplements. Further investigations are warranted to elucidate the mechanisms by which vitamin supplementation can modify breast cancer development. Breast cancer is the most frequently diagnosed cancer globally and also the main cause of death among women . The higet al . In the Shanghai Breast Cancer Study, Dorjgochoo et al [Maruti et al  found nooo et al  evaluateet al [Wu et al  analysedet al [Regarding folic acid supplementation, few data have been published. Data collected from two cohorts , 32 and et al  found anet al [et al [We found 11 studies, six cohort –28, 31, l [et al  found anl [et al , 43, 44.et al [et al [The association between breast cancer risk and vitamin D supplementation has been analysed by three cohort studies , 30, 37.et al  found a l [et al  found a et al [Five cohort –27, 31 aet al [et al [The association between breast cancer risk and multivitamin supplementation has been analysed by six cohort , 32–36 aet al  found anIn the present review, we examined the intake of vitamin supplements and its association with breast cancer risk. Out of the 26 studies, we identified 104 estimates evaluating the effect of vitamin A and beta-carotene, B-group vitamins and folic acid, vitamin C, vitamin D, vitamin E, and multivitamin. We found 14 significant estimates. This review addresses the supposition that the use of multivitamin or single vitamins does not have an effect on breast cancer risk.et al found that supplementations of ten years or longer of multiple vitamins, beta-carotene, vitamin C, and vitamin E were associated with statistically significant reductions for breast cancer risk for postmenopausal women. Moreover, they found a reduced cancer risk also for high intake of beta-carotene in premenopausal women and vitamin E in postmenopausal women. It is known that the oxidative stress and lipid peroxidation are linked to the aetiology of breast cancer, and antioxidants may exert their effect, blocking free radicals, inducing apoptosis in cancer cells, and inhibiting cell proliferation [et al found, in the Women’s Health Initiative Observational Study on postmenopausal women, an inverse association between breast cancer risk and dietary carotenoids in subject with ER+ and PR+, but not with other breast cancer groups [et al can be affected by recall bias inherent in case-control design. Cases might bias their responses to questions on diet after cancer diagnosis. In contrast, cohort and prospective studies have an important advantage because diet assessment is made before diagnosis, and it should be unbiased by cancer experience. The results seem to be in the opposite direction when found in prospective studies  that could modify the estimates.In conclusion, while dietary factors may be crucial in modifying cancer risk, the role of vitamin supplements in preventing breast cancer still remains unclear, considering our review. Although biologic mechanisms exist to support the anticancer effects of vitamins, this review did not address whether supplement use may have an effect on breast cancer risk. Considering the current recommendations and expert report published , 55, theThe authors declare that they have no conflicts of interest.


Estrogen forms a complex with the estrogen receptor (ER) that binds to estrogen response elements (EREs) in the promoter region of estrogen-responsive genes, regulates their transcription, and consequently mediates physiological or tumorigenic effects. Thus, sequence variants in EREs have the potential to affect the estrogen-ER-ERE interaction. In this study, we examined the hypothesis that genetic variations of EREs are associated with breast cancer development.This case-control study involved 815 patients of Asian descent with incident breast cancer and 821 healthy female controls. A total of 13,737 ERE sites in the whole genome predicted by a genome-wide computational algorithm were blasted with single-nucleotide polymorphism (SNP) sequences. Twenty-one SNPs located within 2,000 bp upstream or within introns 1 and 2 of putative genes and with a minor allele frequency greater than 5% were identified and genotyped. Frequencies of SNPs were compared between cases and controls to identify SNPs associated with cancer susceptibility.NRCAM which codes for a cell adhesion molecule, and SNPs of ESR1, the gene coding for ER, on breast cancer risk was found. Interestingly, this combined effect was more significant in women who had experienced a longer period of lifetime estrogen exposure, supporting a hormonal etiology of this SNP in breast tumorigenesis.A significant combined effect of rs12539530, an ERE SNP in intron 2 of ESR1 in determining susceptibility of breast cancer development.Our findings provide support for a role of genetic variation in ERE- The roles of estrogen receptor α (ERα) in initiating tumor development in breast cancer, regulating progression and determining therapeutic protocols and efficacy are well documented -3. AlthoXenopus vitellogenin gene identified a minimal ERE core sequence of 5'-GGTCANNNTGACC-3' [cis-regulatory modules, to predict all possible EREs in the genome. The prediction algorithm of PReMod takes into account the fact that, in higher eukaryotes, cis-regulatory regions often contain several phylogenetically conserved binding sites for different transcription factors [Early work on the TGACC-3' . Since tTGACC-3' ,10. By sTGACC-3' . To distTGACC-3' . In thisTGACC-3' , a new d factors ,14, and ESR1 [This case-control study is part of an ongoing cooperative study aimed at understanding the causes of breast cancer in Taiwan. Breast cancer in Taiwan is characterized by low incidence, early tumor onset, hormone dependency and novel genomic alterations -17. We sESR1 , the genESR1 -21.Experienced research nurses were assigned to administer a structured questionnaire to both patients and controls. The information collected has been described, and the validity of the questionnaire has been addressed and confirmed, in our previous studies -21.At the end of the interview, blood was taken for DNA isolation and genotyping. All samples were examined by laboratory personnel who were blinded to the case/control status of specimens. DNA was extracted from the peripheral blood samples of patients and controls using DNA purification kits .The PReMod database is a genome-wide/transcription factor-wide collection of more than 100,000 computational predicted transcriptional regulatory modules within the human genome ,14. ThesSNPs were genotyped in all samples using Sequenom iPLEX technology . Positive, negative and duplicate controls were included on all plates, with genotypes being autocalled by specialized software  and subsequently confirmed by visual assessment of the data. All assays were performed by individuals blinded to the case-control status of the samples. As a quality control, we repeated the genotyping on 10% of the samples, and all genotype scoring was performed and checked separately by one reviewer who was unaware of the case-control status. The concordance rate for replicate samples was 100%.2 goodness-of-fit test or a Fisher's exact test. The χ2 test for a 2 × 3 contingency table was used to compare genotype frequency between patients and controls. To take account of multiple comparisons, these associations were also assessed using the permutation test provided in the Haploview software tool , run using 10,000 permutations. The association of susceptibility genotypes and breast cancer risk was further evaluated with simultaneous consideration of established risk factors for breast cancer or other significant risk factors in multivariate logistic regression models. Biologic plausibility was the most important criterion for inclusion of variables in the model. Therefore, we included all established risk factors for breast cancer in the statistical models: age, family history of breast cancer, age at menarche, parity and age at first full-term pregnancy (FFTP). Adjusted ORs (aORs) and 95% CIs for genotypes were then estimated.Univariate and multivariate analyses were used to determine the risk factors for breast cancer in this series of study participants, and the odds ratio (OR) and corresponding 95% confidence intervals (95% CIs) were estimated. For individual ERE SNPs, genotype frequencies were assessed for departure from the Hardy-Weinberg equilibrium using either a χESR1 polymorphism in our study participants that we published recently [ESR1-ERE interaction or estrogen-ESR1-ERE interaction in determining breast cancer development. A combination of the joint method and stratified analysis [ESR1 and ERE was associated with breast cancer formation. A joint effect of ESR1-ERE on increased breast cancer risk was explored using conventional logistic regression, a test evaluating whether a statistically significant increase in risk was observed with specific combinations of putative high-risk genotypes in these SNPs (measured by the β estimates from this regression model). In addition, we stratified our study participants on the basis of ESR1 genotype and examined whether breast cancer risk associated with ERE SNPs was particularly significant in specific ESR1 genotype subsets of women. Because we were especially interested in the relationship between the joint effect of ESR1-ERE and breast cancer risk within categories of risk factors representing different levels of estrogen exposure, we performed stratified analysis to test this hypothesis. Therefore, if the identified joint effect of ESR1-ERE SNPs initiated breast cancer by the formation of the estrogen-ER complex, then the relationship between breast cancer risk and the joint effect would not be the same in women who had experienced different lengths of estrogen exposure. This was evaluated by calculating the risk (aOR) of breast cancer associated with the joint effect of ESR1-ERE SNPs in women with a longer or shorter period of total estrogen exposure. For menopausal women, total estrogen exposure was calculated using the formula (age at menopause - age at menarche - years of full-term pregnancy) and, for premenopausal women, age at menopause in this formula was replaced by age at recruitment into this study.We made use of the information on the recently  and explanalysis -21 was aThe risk profile of breast cancer in our study participants was similar to that found in our previous studies -21 and iA total of 13,737 sites  in the whole genome were predicted to be ERE-related sequences using the PReMod algorithm. After blasting these data with online information available from SNP data sets , 322 ER-binding sites were identified as potentially harboring SNPs. One hundred ten of these 322 SNPs were found to contain no variant allele in the Chinese population, resulting in 212 SNPs, 21 of which met our criteria and were genotyped in patients and controls  compared with women carrying the homozygous wild-type genotype, as well as that the carrying of one additional risk allele was associated with a significant increase in risk   or not expressing ESR1 (MDA-MB-231 cells) and examined whether the expression of these putative ER-regulated genes is ER-dependent. The results for NRCAM were more promising, as they showed that this gene could be expressed in an ER-positive breast cancer cell  and between rs12539530 and rs7739506 (P = 0.09) are in line with the suggestion that an interaction between rs12539530 and ESR1 polymorphism is linked to breast cancer risk. To confirm this interaction, we used the joint method and calculated the risk of breast cancer associated with both rs12539530 and ESR1 SNPs using a set of dummy variables representing different combinations of genotypes of ESR1 and rs12539530. In contrast to the increased, but not significant, risk associated with either a high-risk genotype of ESR1 or rs12539530 alone, the greatest risk was found in those patients harboring high-risk genotypes of both rs12539530 and ESR1 SNPs in all women combined  and activator protein 1 (AP-1). After blasting these binding sites with the sequences of 30 SNPs known to reside in this block as reported by HapMap, some SNPs were found to be located within these binding sites, so the second possibility cannot be totally excluded. However, we are prompted to suggest the breast tumorigenic contribution of rs12539530 on the basis of the findings that (1) rs12539530 and ESR1 SNPs jointly increased breast cancer susceptibility and (2) this joint effect was more significant in women with a longer period of estrogen exposure. In addition, this suggestion is mechanistically plausible and is similar to the finding that rs10736303, the SNP generating a putative ERE in intron 2 of FGFR2, has been identified as the most significant gene determining breast cancer susceptibility in recent genome-wide association studies (GWASs) [NRCAM, the gene putatively regulated by rs12539530, has been suggested to be upregulated in ER-positive, but not in ER-negative, breast cancer cell lines [In considering whether our findings represent a true association between this SNP in ERE and breast cancer, the most important issue is the interpretation of the identified association between SNPs and the trait. Because the SNP identified is in an intron, it does not affect amino acid coding and therefore probably does not directly affect protein function. In addition, the observed association between breast cancer and this SNP could be due to the presence of linkage disequilibrium (LD) between this SNP and other SNPs in exons  or in regulatory regions (affecting the expression of these genes). However, the first of these two possibilities is less likely, as we checked the LD block in which rs12539530 is located in the Chinese population in HapMap , which s (GWASs) -31. In all lines , and thiNRCAM is located, has never been reported to be important with regard to breast cancer risk determination in GWASs. This might be partially explained by the relatively low-penetrance effect of SNPs of estrogen-regulating genes. Distinct from those alleles identified by current GWAS, these polymorphic alleles of estrogen-regulating genes would predispose carriers to only a moderately increased risk of developing cancer. Thus, the significance of such SNPs depends not only on their own effect but also on the interaction between the target genes and other functionally related genes  or the promoting effect of reproductive risk factors reflecting estrogen exposure. Our findings are in line with this suggestion.Recent GWASs based on technological advances in high-throughput genotyping have provided information regarding the LD of neighboring polymorphisms. These GWASs have made possible the use of a few hundred thousand SNPs as tags for all other variants, as well as the biobanking of tens of thousands of specimens, allowing their immediate use in whole genome research. These studies have led to the mapping of novel susceptibility loci for many common traits, including breast cancer ,31. InteThis study is a hybrid of candidate gene and genome-wide approaches in which we took advantage of both designs. The well-defined roles of the ER during breast tumorigenesis makes it reasonable to assume that polymorphic genetic variants of EREs, central nodes in the ER pathway, might underlie the variations seen between individuals in their susceptibility to breast cancer. This candidate mechanism lends critical support to the biological plausibility and tumorigenic relevance of our findings. In addition, our genotyping of SNPs on the basis of genome-wide predicted EREs provided a unique opportunity to comprehensively examine putative ERE sites without depending on a prior hypothesis. The successful identification of rs12539530 by using these combined methods suggests that this is a promising approach to identifying the breast tumorigenic contribution of EREs on a genome-wide scale on the basis of information generated by recent technological advances. For example, in our ongoing study, we are exploring the breast tumorigenic role of putative EREs identified using chromatin immunoprecipitation-based methods ,33. The NRCAM, which codes for a cell adhesion molecule, and SNPs of ESR1, the gene coding for ER, on breast cancer risk was found. Our findings provide support for a role of ESR1-ERE polymorphism in determining susceptibility of breast cancer development. This knowledge will be helpful for directing the focus of future experimental studies.Though the promise of personalized medicine, in which the risk and the course of diseases and the efficacy of treatment protocols would be predicted on the basis of a person's genotype, must been tempered with caution, validated molecular tests assessing the patient's germline DNA already drive therapeutic decision-making. On the basis of the well-documented role of ER in breast cancer development and progression, this study explored whether genetic variations in EREs, the sequences bound by ER to activate the transcriptional regulation of target genes, are associated with susceptibility for breast cancer. Notably, the ERE sites genotyped were based on genome-wide prediction, providing a unique opportunity to comprehensively examine putative ERE sites without depending on a prior hypothesis. A significant combined effect of rs12539530, an ERE SNP in intron 2 of CI: confidence interval; ER: estrogen receptor; ERα: estrogen receptor α; ERE: estrogen response element; FFTP: first full-term pregnancy; FTP: full-term pregnancy; GWAS: genome-wide association studies; LD: linkage disequilibrium; OR: odds ratio; SNP: single nucleotide polymorphism.Financial support was provided by grant NSC 98-2628-B-016-001-MY3 (to JCY) from the National Science Council, Taiwan and by grants TSGH-C97-7-S03, DOD-98-28-03 and DOD-98-36 from the Tri-Service General Hospital (to JCY and HMH).The authors declare that they have no competing interests.NRCAM expression in breast cancer cell lines. SD and CWC performed ESR1 genotyping.JCY and CYS participated in the study design and coordination, carried out the participant recruitment and drafted the manuscript. CNH and PEW carried out the genotyping and data analysis. BYB contributed to searching ERE websites using PReMod. HMH, STC and GCH carried out the participant recruitment. WCC and LYH contributed to the validation of Tables S1 and S2. Table S1 presents genotype frequencies of sequence variants of estrogen response element (ERE)-related sequences in breast cancer patients and controls and the adjusted odds ratio (aOR) in relation to breast cancer risk. Table S2 presents breast cancer risk associated with genotypic polymorphism of rs12539530, a single-nucleotide polymorphism (SNP) in the estrogen response element (ERE)-related sequence, stratified by genotypes of ESR1, the estrogen receptor gene.Click here for file