No studies to date have demonstrated a clear association with breast cancer risk and dietary exposure to acrylamide.A 217-item food frequency questionnaire was used to estimate dietary acrylamide intake in 33 731 women aged 35–69 years from the UK Women's Cohort Study followed up for a median of 11 years.μg day−1, 95% CI: 0.98–1.18, Ptrend=0.1). There was a suggestion of a possible weak positive association between dietary acrylamide intake and premenopausal breast cancer after adjustment for potential confounders . There was no suggestion of any association for postmenopausal breast cancer .In all, 1084 incident breast cancers occurred during follow-up. There was no evidence of an overall association between acrylamide intake and breast cancer (hazard ratio=1.08 per 10 There is no evidence of an association between dietary acrylamide intake and breast cancer. A weak association may exist with premenopausal breast cancer, but requires further investigation. Acrylamide is formed principally by the Maillard reaction between the amino acid asparagine and reducing sugars such as glucose . Animal Some evidence has been found of an association between dietary acrylamide and endometrial and ovarian cancers  is participating in the NewGeneris consortium investigating dietary exposures to potentially carcinogenic compounds (The UKWCS is formed from a sample of 35 372 women who completed a 217-item food frequency questionnaire (FFQ) during 1995–1998  occurring after completion of the FFQ. Breast cancer cases’ survival time was taken as date of completing the FFQ until date of diagnosis. Non-cases’ survival was taken as being from date of enrolment until death or end of follow-up (1 January 2008) whichever was the first. Menopausal status was coded using specific criteria related to menstrual and obstetric history and age.−1) or low  reported total energy intake, as were women with prevalent breast cancer. Associations were estimated for premenopausal and postmenopausal women separately and combined, first as a simple model adjusting for age, smoking status and amount smoked (model 1), and second as a full model additionally adjusting for weight, height, physical activity (hours per day sufficiently vigorous to cause sweating), oral contraceptive use, hormone replacement therapy use, parity, age at menarche, alcohol intake (as grams of ethanol per day), energy intake other than from alcohol, and level of education (model 2). Tests for trend were based on fitting the linear trend over the continuous measure of exposure.Cox's proportional hazards regression was used to explore the relationship between estimated dietary acrylamide intakes and risk of breast cancer using Stata version 11 . Women wTo assess sensitivity of results to undiagnosed prevalent breast cancers at the time the FFQ was completed, analyses were repeated after excluding any events occurring within 6 months of completing the FFQ. A further analysis excluding current or ever smokers was undertaken as exposure to acrylamide through tobacco smoke inhalation is likely to be a more important source of exposure than dietary sources alone , with an inter-quartile range of 10–21 μg day−1 (0.15–0.33 μg kg−1 day−1). Only 316 out of 33 061 (1%) of women consumed more than 50 μg day−1 of dietary acrylamide. Only 99 out of 33 061 (0.3%) of participants consumed in excess of 1 μg kg−1 day−1, a threshold of intake cited by the World Health Organization in risk assessment models  . However, there was no such association for postmenopausal breast cancer , with similar numbers of premenopausal and postmenopausal women offering similar power for each comparison.Overall, there was little evidence of an association between dietary acrylamide intake and breast cancer when premenopausal and postmenopausal cancers were combined (hazard ratio=1.08 per 10 end=0.1) . Howeverμg day−1 (inter-quartile range 14–27 μg day−1) or 0.31 μg kg−1 day−1 (inter-quartile range 0.21–0.43 μg kg−1 day−1). However, dose-response trends were very much the same as before.These results were unchanged by excluding women with breast cancers diagnosed within 6 months of completing the FFQ. When these analyses were further restricted to the subgroup of women who had never smoked, the associations between dietary acrylamide and breast cancer were less apparent, though power was much reduced . If halfμg day−1). However, there was some evidence for a positive association with premenopausal breast cancer. There was a weak dose-response relationship, but the association was particularly apparent in the highest fifth of dietary intake (>23 μg day−1).In our large UK cohort, we have shown no association between dietary acrylamide intake and increased risk of breast cancer, either overall or for postmenopausal breast cancer, for modest intakes consumed as part of a usual diet  is in keeping with figures from the Netherlands and US , but verEven in our relatively health-conscious cohort of middle-aged women, the main sources of acrylamide were chips and crisps. It is therefore possible that any measure of dietary acrylamide intake may be a proxy for poor diet, which could predict breast cancer. This in turn may be predictive of increased risk of breast cancer through other dietary risk factors such as higher fat or processed meat intakes (As this is a prospective study, recall bias is unlikely. However, it is possible that current estimates of acrylamide content in food are incomplete, which may have lead to some misclassification of exposure categories. In addition, estimates of acrylamide exposure were based on the responses to one FFQ administered at recruitment into the cohort. As the consumption of foods thought to be contaminated with acrylamide may have changed over time, it is possible that our estimates of exposure might have been improved with repeated dietary assessments. The range of acrylamide levels reported for specific products, with different methods of preparation, is wide, and even within brands considerable variation occurs between different batches . As the The strongest suggestion of a positive association comes from a large prospective cohort (There is some mechanistic evidence to support this epidemiological evidence. In animal studies, acrylamide is causally associated with mammary tumour development (Given the widespread presence of acrylamide in food and the relatively common nature of breast cancer, even a small association between the two would have important public health implications. In this first UK study, no overall association between dietary acrylamide intake and breast cancer risk was found, though a weak positive association was observed among premenopausal breast cancers, but further investigation is needed.


XRCC1 gene is involved in the base excision repair pathway. We assessed the associations of polymorphisms and haplotypes in XRCC1 with skin cancer risk in a nested case–control study within the Nurses' Health Study  and 300 basal cell carcinoma (BCC), and 873 controls). We genotyped four haplotype-tagging single-nucleotide polymorphisms . There was no significant difference in frequency distribution between cases and controls for any of the five inferred common haplotypes. We observed that the 399Gln allele was inversely associated with SCC risk. This inverse association was only seen among those who had five or more lifetime sunburns, those with a family history of skin cancer, and those in the highest tertile of cumulative sun exposure in a bathing suit, but not among those with low risk defined by these risk factors. We also observed a significant association of the carriage of 194Trp allele with increased SCC risk, which was modified by family history of skin cancer. These two polymorphisms were not associated with BCC or melanoma risk. Our data suggest that the Arg194Trp and Arg399Gln polymorphisms may be differently associated with skin cancer risk according to exposure dose and skin cancer type.The  The genotoxic effect of sunlight exposure has been demonstrated in the aetiology of both melanoma and nonmelanocytic skin cancer .β , and squamous cell carcinoma (SCC)) in a nested case–control study within the Nurses' Health Study. We further investigated the gene–environment interactions between XRCC1 genetic variation and sun exposure-related risk factors on skin cancer risk.Base excision repair (BER) is responsible for repair of oxidative DNA damage and single-strand breaks . It has β . This suin situ cases) up to June 1, 2000 with no previously diagnosed skin cancer. All available pathologically confirmed melanoma and SCC cases and 300 self-reported BCC cases randomly selected from about 2600 available self-reported BCC cases were included. The validity of self-report of BCC is high in this medically sophisticated population (90%) (in situ cases), 286 SCC cases, 300 BCC cases, and 874 matched controls. Owing to the absence of African-American cases, one African-American control was excluded to avoid potential population stratification. We mailed to 758 living cases and 804 living controls a supplementary questionnaire on lifetime sun exposure and other skin cancer risk factors. In all, 695 cases responded, 15 cases refused to participate, and 48 cases did not respond after three mailings (participation rate=92%). Among controls, 713 responded, nine refused, and 82 did not respond (participation rate=89%). The study protocol was approved by the Committee on Use of Human Subjects of the Brigham and Women's Hospital, Boston, MA, USA.The Nurses' Health Study was established in 1976, when 121 700 females registered nurses between the ages of 30 and 55, residing in 11 larger US states, completed a self-administered questionnaire on their medical histories and baseline health-related exposures. Updated information has been obtained by questionnaires every 2 years. Between 1989 and 1990, blood samples were collected from 32 826 volunteers of the cohort members. Eligible cases in this study consisted of women with incident skin cancer from the subcohort who gave a blood specimen, including SCC and BCC cases with a diagnosis anytime after blood collection up to June 1, 1998 and melanoma cases  pigmentation, constitutional, and susceptibility factors; (2) history of residence (states and towns), sun exposure habits, and severe sunburns at different ages; and (3) family history of skin cancer . In addition, the 11 states of residence of cohort members at baseline were grouped into three regions: Northeast , Northcentral (MI and OH), and West and South . The reliability of the measurements obtained on both the prospective and retrospective data  was approximately in the same magnitude in terms of correlation coefficient and mean change among the cases and controls and the odds ratios (ORs) based on both questionnaires were similar, indicating that the retrospective assessment was not likely to substantially bias the estimates of risk in this study, at least for these variables.In order to estimate sunlight exposure for each subject, a UV database for 50 US states was developed. The database used reports from the Climatic Atlas of the US, which reported mean daily solar radiation (in Langleys) at the earth's surface for weather stations around the country. The records of average annual solar radiation for January and July were extracted to represent winter and summer radiation, respectively. The mean solar radiation for each state was derived from the average of UV values measured in weather stations within that state, and both summer and winter radiation indices were developed for each state. A cumulative lifetime sun exposure was developed by combining the UV database and the information obtained from the supplementary questionnaires. Questions about sun exposure while wearing a bathing suit were used to define a cumulative lifetime sun exposure variable for this behavior.XRCC1 gene, using data derived from the resequencing of the exons and adjacent intronic and noncoding regions of the gene in a multiple-ethnicity group of 90 samples from the NIH DNA Polymorphism Discovery Resource available from the Coriell Institute for Medical Research (As described previously (Genotyping was performed by the 5′ nuclease assay (TaqMan®), using the ABI PRISM 7900HT Sequence Detection System , in 384-well format. TaqMan® primers and probes were designed using the Primer Express® Oligo Design software v2.0 (ABI PRISM). Laboratory personnel were blinded to case–control status and blinded quality control samples were inserted to validate genotyping procedures; concordance for the blinded samples was 100%. Primers, probes, and conditions for genotyping assays are available upon request.χ2 test to assess whether the XRCC1 genotypes were in Hardy–Weinberg equilibrium and to determine P-values for differences in haplotype frequencies between cases of each type and common controls. We used a common control series in data analysis to increase the statistical power. Unconditional logistic regression was employed to calculate OR and 95% confidence interval (CI) to assess the risks of the three types of skin cancer for XRCC1 genotypes. A test for trend was calculated across the three genotypes for each polymorphism. The results did not change substantially in the matched analysis for each of the three types of skin cancer with their own matched controls. We used a likelihood ratio test (LRT) to evaluate heterogeneity in the effects of the XRCC1 genotypes on different types of skin cancer in polytomous logistic regression models . We also modelled XRCC1 genotypes as ordinal variables and environmental exposures as continuous variables to assess the statistical significance of interactions by testing the significance of a single multiplicative interaction term . All P-values were two-sided.To test the statistical significance of interactions between the environmental exposures and the vs controls: Caucasian, 85.3 vs 85.7%; others, 11.7 vs 11.6%; one Asian melanoma case and one control; one Hispanic SCC case and two controls). Significantly positive associations were observed of lighter natural skin color, lighter natural hair colour, child or adolescent tendency to burn, and the number of palpably raised moles on arms with the risk of all three types of skin cancer. The risk for the highest tertile of the susceptibility score was about three-fold higher for SCC and BCC and four-fold higher for melanoma, compared to the lowest tertile. A family history of skin cancer was a risk factor for the three types of skin cancer. Cases of each skin cancer type were more likely to have used sunlamps or attended tanning salons. The number of lifetime severe sunburns that blistered was significantly associated with all three types of skin cancer. Women in the West and South regions were more likely to be diagnosed with SCC or BCC compared to those in Northeast.The mean ages at diagnosis of melanoma, SCC, and BCC cases were 63.4, 64.7, and 64.0 years, respectively, and that of controls, 64.5 years. Self-reported major ethnicity was similar between cases and controls . Our data showed that the two variants 194Trp and 399Gln were not in the same common haplotype. Carriage of either variant allele was not associated with altered skin cancer risk compared with being homozygous wild type for both alleles (data not shown). No overall associations between C26602T and Gln632Gln and three types of skin cancer risks were found. There was no significant heterogeneity in the main effect of each genotype on the three types of skin cancer (http://egp.gs.washington.edu).We therefore evaluated the four haplotype-tagging SNPs in relation to skin cancer risk , and this excess risk was attenuated among those who carried the Gln allele. The Gln allele was significantly inversely associated with SCC risk among those who had five or more severe sunburns in their lifetime , but positively associated with SCC risk among women who never had severe sunburns . This interaction was statistically significant . The 399Gln allele was not significantly associated with lifetime sunburns among controls. The interaction pattern described above was limited to those in the West and South regions, but not among those in Northeast and Northcentral regions (data not shown).We evaluated potential interactions between the Arg399Gln polymorphism and lifetime severe sunburns, family history of skin cancer, cumulative sun exposure with a bathing suit, and geographic region on SCC risk  than women without family history of skin cancer   among controls.We observed that the inverse association of n, 0.02) . It was P for trend, 0.02), whereas no apparent association between the 399Gln allele and SCC risk was observed among women in the lowest or intermediate tertile . Compared with 194Trp noncarriers without family history, the 194Trp carriers with a family history of skin cancer had a multivariate OR of 3.45 , whereas the multivariate OR was 1.16  for 194Trp carriers without family history, and 1.19  for 194Trp noncarriers with family history. No interactions were observed between the Arg194Trp and other risk factors on SCC risk. We did not observe any notable interactions between the two polymorphisms Arg399Gln and Arg194Trp and the above risk factors on BCC or melanoma risk.We also observed a significant interaction between the Arg194Trp polymorphism and family history of skin cancer on SCC risk  and SCC . However, these authors observed an inverse association of the 399Gln allele among women with two or less sunburns, but not among those with three or more sunburns. We did not observe this interaction pattern either in the whole study or among women in the Northeast region.One previously published study assessed the Arg399Gln polymorphism and nonmelanoma skin cancer risk in a population-based case–control study in New Hampshire (XRCC1 194Trp allele was significantly associated with increased risk of SCC. Given the consistent relation of the 194Trp allele with the reduced cancer risk of bladder, lung, breast, and stomach observed in some case–control studies (We observed that the XRCC1 Arg399Gln polymorphism with skin cancer risk may vary according to exposure dose and cancer type. We observed some evidence that the 399Gln allele was related to a decreased risk of SCC, but not that of melanoma or BCC. The effect modification of the 399Gln allele on SCC risk associated with sunburn suggests that oxidative DNA damage by the UVA predominantly present in sunlight may play an important role in the development of SCC. Given the number of comparisons, the findings should be interpreted with caution and confirmed by other studies.Our data suggest that the association of the


Included were 608 incident breast cancer cases and 609 age frequency matched controls, aged 30–80 years old. Cases and controls were interviewed by trained study interviewers, using a standardized, structured questionnaire, to obtain information on lactation and other major risk factors. Parous women who reported ever lactation had a borderline significantly reduced risk of breast cancer . An OR of 0.53  was observed in those having breastfed more than 3 children compared to those who never lactated. Women having breastfed their first child for more than 13 months had an OR of 0.47  compared to those who never breastfed. Lifetime duration of lactation also showed a risk reduction while none of the ORs were statistically significant. Further stratification by menopausal status showed a risk reduction related to lactation for both pre- and postmenopausal women, while the relationship is less consistent for the latter. These results support an inverse association between breastfeeding and breast cancer risk. © 2001 Cancer Research Campaign


We conducted a cohort study with a nested case–control analysis to evaluate the effect of anti-inflammatory drugs in breast cancer incidence using the General Practice Research Database. Women taking aspirin and paracetamol for 1 year or longer had an odds ratio (OR) of 0.77 : 0.62,0.95) and 0.76 , respectively, compared to nonusers. Daily doses of aspirin (75 mg) and paracetamol (up to 2000 mg) showed the greatest reduced risk. Use of non-aspirin nonsteroidal anti-inflammatory drugs for more than 1 year was not associated with a reduced risk of breast cancer , and the corresponding estimate among users with at least 2 years duration was similar. Our findings suggest that aspirin at cardioprophylactic doses as well as paracetamol at analgesic doses is associated with a reduced risk of breast cancer. There is mounting epidemiological evidence suggesting that nonsteroidal anti-inflammatory drugs (NSAIDs) may substantially reduce the risk of colorectal cancer . The effin vitro studies suggest that glucocorticoids have a direct inhibitory effect on proliferation of mammary cancer cells .We used data from the GPRD. This database contains computerised information entered by general practitioners (GPs) in the UK . Data onWe identified all female subjects 30–79 years old between January 1995 and December 2001. Women became members of the study population on the first day of the study period when they met the criteria of at least 1 year enrolment with the GP and 1 year since the first computerised prescription. That date was their start date. Study members with a code for cancer before start date were excluded. We also excluded women 70 years and older at start date with a follow-up greater than 1 year and no data recorded during their total follow-up time: an indicator of nonassistance with their GP. Our final study cohort comprised 734 899 women.All study members were followed from start date until the earliest occurrence of one of the following end points: recorded diagnosis of breast cancer, any cancer other than breast, age of 80 years, death or end of study period (December 2001).We identified 4005 patients with a code of breast cancer and manually reviewed their computerised patient profiles. Information included demographic data and all clinical information with no personal identifiers. We excluded 297 women: the main reasons were the computer diagnosis was subsequently not confirmed (60%) and prevalent cases (31%). A previous study validating a large number of cancer cases documented a high reliability of cancer diagnoses recorded in the GPRD (n=3708) identified in the study cohort were used in the nested case–control analysis and we considered their date of initial diagnosis as index date. A date during the study period was generated at random for every member of the study cohort. If the random date of a study member was included in her eligible person-time, we used her random date as the index date and marked that woman as an eligible control. This selection mechanism allows that the likelihood of being selected as a control is proportional to the person-time at risk. The same eligibility criteria were applied to controls as to cases. In total, 20 000 controls were frequency-matched by age  and calendar year from the list of all eligible controls.All cases of breast cancer (Estimates of odds ratio (OR), assumed to be valid estimates of the relative risk, and 95 percent confidence interval (95% CI) associated with use of aspirin, non-aspirin NSAIDs, paracetamol, and oral steroids compared to nonuse were computed using unconditional logistic regression. We ascertained patients with previous benign breast disease recorded at least more than 1 year before the index date. Other risk factors like alcohol intake, body mass index (BMI) and hormone replacement therapy (HRT) were also ascertained based on their medical history recorded in the GPRD. We also elicited subjects’ use of health services  in the 2 years prior to the index date. All estimates of OR were adjusted for age, calendar year, BMI, alcohol intake, smoking status, HRT use and prior benign breast disease.We defined three time windows of exposure for aspirin, non-aspirin NSAIDs, paracetamol and oral steroids: current use, past use and nonuse. Current use was defined as use that lasted until the index date or ended in the year prior to the index date based on the supply of drug therapy as prescribed by the GP. Past use was use that ended more than 1 year before the index date. Finally, the time window of nonuse was defined as no recorded use at any time before the index date. Current users were subdivided into short-term and long-term users. Short-term users were patients whose duration of treatment was shorter than 1 year and long-term users were patients treated for a period longer than 1 year.The effect of dose was also analysed among long-term users. Aspirin daily dose was assigned into three groups: 75, 150 and 300 mg. Specific cutoff values for each individual non-aspirin NSAID were used to group non-aspirin NSAID daily dose into two categories; low-medium and high dose. Daily doses of paracetamol up to 1000 mg were assigned into a low-dose category, doses above 1000 mg and up to 2000 mg were assigned into a medium dose category, while doses above 2000 mg were grouped into a high-dose category. Oral steroid daily doses up to 10 mg of prednisolone  were classified as low-medium, while daily doses of 10 mg and above were classified as high. An additional analysis using 2 years time lag (advancing the index date by 2 years in cases and controls) was performed.The incidence rate of breast cancer in our study population was 156 per 100 000 person-years among women 30–79 years old, well in line with other reports from the UK .Aspirin was associated with a decreased risk of breast cancer  compared to nonusers. Women with the longest duration of use (4 years or more) presented a lower risk  .We found that use of aspirin and paracetamol was associated with a reduced risk of breast cancer of about 20%. On the other hand, we found little evidence for a protective effect of non-aspirin NSAIDs. These results are quite similar to those found in a recent study performed with an identical design and using the same source population that addressed the risk of another hormone-dependent cancer in men, in this case prostate cancer .We identified 10 previous studies evaluating the association between aspirin and breast cancer. Among them, eight reported an estimate in line with our observed 20% risk reduction  use of NSAIDs, paracetamol or aspirin is not recorded on computer files. However, the proportion of OTC use  out of the total use of chronic treatment is small. Yet, nondifferential misclassification of drug use due to noncompliance or OTC use would tend to attenuate the true association between the specific drug group and breast cancer  did not materially change the results (data not shown).Although we recorded information regarding traditional risk factors for breast cancer including age, alcohol use, smoking, BMI, HRT use and previous breast abnormalities, there is still room for some confounding from either measurement bias or other risk factors that we were not able to elicit such as age at menarche, parity, family history, age at first child or germline mutation. However, we would expect these risk factors to be rather evenly distributed among users and nonusers of the study drugs and consequently affecting very little, if any, our estimates of effect.In summary, our findings suggest that use of aspirin at cardioprophylactic doses is associated with a reduced risk of breast cancer, but there was little evidence for a reduced risk among users of non-aspirin NSAIDs. We also found long-term use of paracetamol to be linked to a reduced risk of breast cancer.


Mean trait anxiety scores were higher than general population data  but not significantly different from published data from other screening samples. Overestimators  and underestimators  reported significantly different risk estimates (i.e. increased accuracy) after counselling, but significant inaccuracies persisted. Over- (n = 12) and underestimators (n = 60) were still inaccurate in their risk estimates by a factor of 2 after counselling. Thirty per cent of the sample scored above the cut-off (5/6) for case identification on a screening measure for psychological distress, the General Health Questionnaire (GHQ). GHQ scores were significantly lower after counselling  with no evidence of increasing risk estimate causing increased distress. The risk of distress after counselling was greater for younger women and those who were more distressed at first presentation. The counselling offered was effective in increasing the accuracy of risk perceptions without causing distress to those who initially underestimated their risk. It is worrying that inaccuracies persisted, particularly as the demand for service has since reduced the consultation time offered in this clinic. Further work is needed to evaluate alternative models of service delivery using more sophisticated methods of assessing understanding of risk. © 1999 Cancer Research CampaignWomen referred to a familial breast cancer clinic completed questionnaires before and after counselling and at annual follow-up to assess their risk estimate and psychological characteristics. The aims were to determine whether those who attended the clinic overestimated their risk or were highly anxious and whether counselling influenced risk estimates and levels of distress. Women (


The relation between hormonal and lifestyle factors and breast cancer risk in nulliparae was investigated using data from two case-control studies conducted in Italy between 1983 and 1994. The study included 1041 nulliparae with histologically confirmed incident breast cancer and 1002 nulliparous controls admitted to hospital for a wide range of acute, non-neoplastic, nonhormone-related diseases. In premenopausal nulliparae, there was an inverse relation with age at menarche , while no association emerged in postmenopausal. Breast cancer risk increased with age at menopause, the OR being 1.91 (95% CI 1.26–2.90) for nulliparae reporting age at menopause ≥ 53 years compared with < 45. Abortion was not related to breast cancer risk, the OR being 0.92 for any spontaneous, 0.97 for any induced and 0.77 for ≥ 2 total abortions compared to none. The OR was 1.75 (95% CI 1.03–2.97) for women reporting their first abortion at age ≥ 30 years compared with < 30. Oral contraceptives and hormone replacement therapy in menopause were moderately related to risk. The OR was 2.71 (95% CI 1.85–3.95) in nulliparae with a family history of breast cancer and 1.60 (95% CI 1.20–2.14) in those with a history of benign breast disease. Compared with nulliparae reporting a low physical activity, the OR was 0.79 (95% CI 0.54–1.16) for those reporting intermediate/high activity. Breast cancer risk increased with total energy intake, the OR being 1.65 (95% CI 0.99–2.75) in the highest tertile; beta-carotene was inversely related to risk  for the highest tertile. Thus, most risk factors for breast cancer in nulliparae were similar to those in women generally. © 1999 Cancer Research Campaign


Results confirmed our earlier findings. Based on self-reports, former college athletes had a significantly lower risk of breast cancer than the non-athletes. The OR for the 15-year incidence of breast cancer is 0.605 with 95% confidence interval (CI) (0.438–0.835); the 15-year incident breast cancers were 64 among the athletes and 111 among the non-athletes. Among women under 45 the protective effect of physical activity on the risk of breast cancer is considerably greater; odds ratio (OR) = 0.164, 95% CI (0.042–0.636). Athletic activity during the college and pre-college years is protective against breast cancer throughout the life span, and more markedly among women under 45. These results confirm our earlier findings and the findings of other investigators. © 2000 Cancer Research CampaignA growing body of evidence indicates that physical activity is protective against breast cancer. In 1996–97, we conducted a 15-year follow-up of 5398 college alumnae comprised of former college athletes with their non-athletic classmates. Participants completed a detailed mailed questionnaire on their health history from 1981–82 to the present. Excluding women who had died and non-deliverable questionnaires, 84.7% (


Mammographic density is a strong risk factor for breast cancer. Our objective was to examine its association with polymorphisms identifying breast cancer susceptibility loci that were ascertained in recent genome-wide association studies.FGFR2, rs3803662 and rs12443621in TOX3, rs3817198 in LSP1, rs981782 and rs10941679 near HCN1/MRPS30, rs889312 in MAP3K1, and rs13387042 at 2q) were examined. Mammographic density was quantified with a computer-assisted method as the percent dense area: the area of radiologically dense fibroglandular tissue relative to the total breast area that also includes radiologically lucent fatty tissue.Subjects were 825 women who participated in previous case–control studies of mammographic density and genetic factors nested within the Multiethnic Cohort study and were from three ethnic groups . Eight polymorphisms  had 3% to 4% higher densities than women with two A alleles. The polymorphism rs10941679 near HCN1/MRPS30 was also associated with percent dense area; women who were homozygous for the G allele (previously associated with increased breast cancer risk) had 4% to 5% lower densities than women with at least one A allele. The other polymorphisms were not associated with percent dense area.The polymorphism rs12443621 in TOX3 influences breast cancer risk.The available data suggest that the effects of most of these polymorphisms on breast cancer are not mediated by mammographic density. Some effects may have been too small to be detected. The association with rs12443621 may provide clues as to how variation in  Mammographic density is related to breast cancer; women within the highest categories of mammographic density are at four to six times higher risk than women within the lowest categories . MammogrMammographic density is influenced by genetics -4, with FGFR2, TOX3 (formerly known as TNRC9), LSP1, HCN1/MRPS30, MAP3K1 and at 2q that had not previously been considered [FGFR2 [TOX3 [MAP3K1 [LSP1 [HCN1/MRPS30 [Genome-wide association studies (GWAS) have identified breast cancer susceptibility loci in or near genes such as nsidered -10. Two nsidered ,12. Bothd [FGFR2 , rs38036K1 [LSP1 ,12), or 1/MRPS30 ).Subjects were the subset of women who were included in a case–control study of mammographic density ,14 and wThe mammographic density study included MEC participants from Hawaii, and included 607 cases who were diagnosed with invasive breast cancer by the end of December 2000 and 667 controls who were frequency matched to cases in ethnic and 5-year age groups . ParticiTo enable case–control studies of genetic susceptibility, a blood sample was requested during 1995 to 2000 from breast cancer cases occurring in the MEC and from a random sample of controls, and subsequently from a substantial portion of the MEC participants . These pThe current analysis involved 361 cases and 464 controls who were part of the mammographic density study, had at least one polymorphism of interest genotyped for the replication steps of the GWAS, and were in the major ethnic groups represented in Hawaii . The MEC and its genetic substudies were approved by the Institutional Review Boards at the University of Southern California and the University of Hawaii. The mammographic density study was approved at the University of Hawaii. Participants provided informed consent for both case–control studies.Mammograms were available for a mean of 2.8 different dates per woman. Mammograms were performed before diagnosis for all but five cases. For these cases, a mammogram was available only at the time of diagnosis so the film of the contralateral breast was used. Otherwise, if mammograms from both breasts were available on a single date, their measures were averaged.Films of the craniocaudal view were digitized using a Lumisys 85 scanner  that creates images with pixel size equal to 260 μm and has an optical density from 0 to 4.1. Mammographic density was determined using the computer-assisted software, Cumulus, developed at the University of Toronto, Canada . This soBlood was obtained from subjects at their homes, processed within 8 hours, and stored at -80°C. DNA was purified from white blood cell fractions using the QIAamp 96 DNA Blood Kit . The details of the 5'-nuclease assay (Taqman)  have beeP < 0.01) only occurred for the rs981782 genotypes among Native Hawaiian subjects (P = 0.0008), possibly due to instability arising from small numbers (n = 39).Chi-square test statistics for the hypothesis of Hardy–Weinberg equilibrium and the corresponding levels of significance were calculated by ethnic group. Significant deviation from what was expected , with any risk allele , or with both risk alleles  were performed. Adjustment was made for age and the square of age at the time of each mammogram because percent dense area decreases at a rate that slows over time [P values are two-sided with the level of significance set at P < 0.05. SAS 9.1  was used for the analysis.To investigate the associations between the genotypes and percent dense area, mixed models were applied to account for subjects with multiple mammographic readings over time. This method is nearly equivalent to modeling each woman's average percent dense area over time except that it uses maximum likelihood rather than ordinary least squares estimation. The adjusted mean percent dense area was estimated by genotype. Tests for the difference in mean percent dense area and associated ver time , for ethver time . Variablver time ,11,21,222. Women were of White (32%), Japanese American (44%) or Native Hawaiian (24%) ethnicity. At the time of the first mammogram used in this study, 201 (24%) of the women were premenopausal; because subjects had multiple mammographic readings over time, a lower proportion (18%) of mammograms used was taken when the subjects were premenopausal. The mean percent dense area was 32.7 ± 22.2.For each polymorphism of interest, 634 to 806 subjects were genotyped; 825 subjects had at least one of the polymorphisms genotyped. Characteristics of these 825 subjects are presented in Table TOX3 was associated with a significantly higher percent dense area than being homozygous for the A allele (Pd = 0.03)  and seemed to be more pronounced in premenopause. This association was nonsignificantly stronger in cases than in controls (Pinteraction = 0.054); with dense area, however, the association was significantly stronger in cases and no association was evident in controls (Pinteraction = 0.01). Furthermore, among Whites homozygotes of the G allele had a higher absolute dense area, among Native Hawaiians homozygotes of the G allele had a lower absolute dense area, and no association was observed among Japanese Americans (Pinteraction = 0.002).Having any G allele of rs12443621 in HCN1/MRPS30 was associated with a significantly lower percent dense area than being having any A allele (Pr = 0.02) ; with dense area, the association was significantly stronger in cases (Pinteraction = 0.02).Being homozygous for the G allele of rs10941679 near Pinteraction > 0.05). Menopausal status with HRT significantly modified some associations, suggesting that, in users of combined HRT, being homozygous for the T allele of rs2981582 in FGFR2, being homozygous for the T allele of rs981782 in HCN1, or being homozygous for the A allele of rs13387042 at 2q35 was associated with decreased dense area , and each C allele of rs3817198 in LSP1 was associated with increased percent dense area (Pinteraction = 0.003).Percent dense area and dense area were not significantly associated with any other polymorphism overall and no other interactions were detected with either case status or ethnicity (TOX3 had 3% to 4% higher percent dense area than women homozygous for the A allele; the association was stronger in premenopausal women and breast cancer cases. Previously, each G allele has been found to be associated with increased breast cancer risk [HCN1 and MRPS30 was also associated with percent dense area; women who were homozygous for the G allele had 4% to 5% lower densities than women with at least one A allele. The G allele, however, has been found to be associated with increased breast cancer risk in a previous study under a co-dominant model [LSP1 was associated with higher percent dense area in subgroups of their samples: in premenopausal women [The rationale for this investigation was that breast cancer susceptibility loci may also be related to another strong risk factor, mammographic density. Two polymorphisms examined were associated with percent dense area. Women with any G allele of rs12443621 in cer risk , whereascer risk . The polnt model . This pont model ,12. All al women , in sterFGFR2; each copy of the T allele of rs2981582 was found to be associated with a 26% increased breast cancer risk [FGFR2 would be related to mammographic density. In this study, however, percent dense area was nonsignificantly decreased with the number of risk alleles.The polymorphisms examined in the present study were identified in GWAS of breast cancer ,9. The scer risk  and becaTOX3, LSP1, HCN1, MAP3K1 and at 2q35 are less strongly associated with breast cancer risk than polymorphisms in FGFR2; the increased risks range from 4% to 20% with each risk allele [TOX3 contains a high mobility group box motif that suggests it is a transcription factor [TOX3 affects breast cancer risk is unknown. Our results and those of others [TOX3 to increase both mammographic density and breast cancer risk. HCN1 is adjacent to other genes that may be related to breast cancer risk including FGF10 and MRPS30 [LSP1, lymphocyte-specific protein 1, may be involved in wound healing [FGFR2, TOX3, HCN1, and at 2q35 may be more strongly associated with estrogen receptor-positive cancers [Polymorphisms in or near k allele ,16,24,25n factor , but thef others  suggest  healing , which i healing ,28. Some cancers ,16,25, b cancers .A limitation of the current study was that the sample size was insufficient to detect interactions with case status, ethnicity, and menopausal status with HRT or to detect small effects. With the sample size in this study, a linear increase per allele of 2% to 3% percent dense area could be detected with 80% power. If an allele increases breast cancer risk between 4% and 26% ,16 and iSelection bias was unlikely to mask associations. Although our sample was a subset of the MEC with a slightly different age and ethnic distribution, we controlled for these factors in the analysis; other unknown or unmeasured factors that influenced selection into this analysis are probably less likely than age or ethnicity to be related to either genotypes or mammographic density. The association between the polymorphisms and mammographic density is therefore unlikely to have been modified by these factors.TOX3 was associated with percent dense area under a dominant model. Some biologic plausibility exists for associations with the other polymorphisms examined but possibly the associations were too small to be detected. The available data, however, suggest that the effects of most of these polymorphisms on breast cancer are not mediated by mammographic density. The association of mammographic density with rs12443621, if confirmed, may provide clues as to how variation in TOX3 influences breast cancer risk.The rs12443621 polymorphism in GWAS: genome-wide association study; HRT: hormone replacement therapy; MEC: Multiethnic Cohort.The authors declare that they have no competing interests.CGW established the concept of this project, carried out the statistical analysis, interpreted the results, and wrote the manuscript. GM established the mammographic density study and provided input into the interpretation of the data. CAH provided input into the study concept and the interpretation of results, and was involved in the parent genetic susceptibility studies. MV provided input into the study concept, statistical analysis, and interpretation of results. ISP performed initial data manipulation in the mammographic density study and provided expertise in the statistical analysis. LLM provided input into the study concept, statistical analysis and interpretation of results, and was involved in the parent genetic susceptibility studies. BEH and LNK are the Principal Investigators of the MEC. Furthermore, all authors provided feedback on the initial draft of the manuscript and approved the final manuscript.An Adobe file containing the table "Adjusted mean percent dense area and dense area by genotype."Click here for file


Despite the endocrine system activity exhibited by polychlorinated biphenyls (PCBs), recent studies have shown little association between PCB exposure and breast cancer mortality.To further evaluate the relation between PCB exposure and breast cancer risk, we studied incidence, a more sensitive end point than mortality, in an occupational cohort.We followed 5,752 women employed for at least 1 year in one of three capacitor manufacturing facilities, identifying cases from questionnaires, cancer registries, and death certificates through 1998. We collected lifestyle and reproductive information via questionnaire from participants or next of kin and used semiquantitative job-exposure matrices for inhalation and dermal exposures combined. We generated standardized incidence ratios (SIRs) and standardized rate ratios and used Cox proportional hazards regression models to evaluate potential confounders and effect modifiers.n = 257), and regression modeling showed little effect of employment duration or cumulative exposure. However, for the 362 women of questionnaire-identified races other than white, we observed positive, statistically significant associations with employment duration and cumulative exposure; only smoking, birth cohort, and self- or proxy questionnaire completion had statistically significant explanatory power when added to models with exposure metrics.Overall, the breast cancer SIR was 0.81 (95% confidence interval, 0.72–0.92; We found no overall elevation in breast cancer risk after occupational exposure to PCBs. However, the exposure-related risk elevations seen among nonwhite workers, although of limited interpretability given the small number of cases, warrant further investigation, because the usual reproductive risk factors accounted for little of the increased risk. The increase in breast cancer rates in recent decades has prompted researchers to explore the role of environmental factors in breast cancer etiology. One such factor is exposure to polychlorinated biphenyls (PCBs), which were manufactured in the United States for a variety of commercial uses from the 1930s until the late 1970s, when they were banned. These environmentally persistent compounds have exhibited endocrine system activity in the laboratory  and thusCYP1A1 polymorphisms  in two plants that produced electrical capacitors in New York State and Massachusetts were studied by No previous study of these capacitor worker cohorts found large or significant elevations for female breast cancer mortality, and studies that examined exposure–response trends by duration  or exposThe study population consisted of women included in prior retrospective cohort mortality studies of workers at three capacitor manufacturing plants in the United States. Plants 1 and 2, located in New York State and Massachusetts, respectively, employed a total of 13,321 women. Plant 3, located in Indiana, had 857 female employees. From the combined base cohort of 14,178 potentially PCB-exposed women, we restricted the present study to the 5,752 women who worked for at least 1 year at any of the three capacitor plants . The minimum employment restriction was motivated by the difficulty in locating women with short-term employment for questionnaire follow-up. Follow-up for breast cancer incidence began after 1 year of employment. This project received approval from the National Institute for Occupational Safety and Health (NIOSH) Human Subjects Review Board (HSRB). Per the HSRB-approved protocol, a returned questionnaire was considered implied informed consent. For study participants interviewed by telephone, we obtained oral informed consent before the interview.a) incident breast and other cancer diagnoses and details necessary to confirm diagnoses via medical records; b) relevant nonoccupational risk factors for breast cancer, such as family history and reproductive history; and c) demographic and lifestyle information such as ethnicity, education, height, weight, and smoking history. Because we knew that plant 2 included a number of workers of Cape Verdean origin, we offered Spanish- and Portuguese-language translations of the questionnaire to non-English-speaking respondents upon request. Questionnaires were collected from 1998 through 2000. Women reporting breast cancer diagnoses after 1998 were not considered cases.We mailed a self-administered questionnaire to all women or their next of kin for whom we could determine valid addresses. After two mailings and a reminder postcard, we telephoned nonrespondents to obtain the information. We identified addresses and telephone numbers using sources such as the Internal Revenue Service (IRS), the U.S. Postal Service, motor vehicle registration records, credit bureaus, and telephone number lookup services. The questionnaire collected information on Reliable information on race was generally not available from plant records. For many deceased workers, we obtained race information from death certificates and other vital status sources. The questionnaire actively sought race information, asking women to select from the following categories: white, black, American Indian or Alaskan Native, Asian or Pacific Islander, and other. Respondents selecting “other” for race were prompted to further specify their race. The questionnaire asked separately whether the worker was of Hispanic origin.For the present study, we used race data from the plants and from death certificates only for calculating preliminary standardized incidence ratios (SIRs) and standardized rate ratios (SRRs). In these analyses, we assumed workers of unknown race to be white. We limited further analyses evaluating the effects of race to the subcohort with questionnaire data and used only the self- or proxy-reported data to classify workers by race. This approach differs from that used in previous studies of these cohorts , which uInternational Classification of Diseases (ICD) revision in effect at time of death . After we obtained the most current addresses, we evaluated the distribution by state of address and decided to also match our file against the Florida registry (covering 1981–1998) and the California registry (covering 1988–1998). Matching is done by staff at the registries, and data are not available to the public.n = 10). However, we included breast cancer cases reported in medical or cancer registry records (even if unreported by a woman or her next of kin) in the analysis (n = 82). We also included breast cancer cases for which we obtained no medical record. Medical record or cancer registry confirmation of the diagnosis was available for 43% and 55% of cases, respectively.Questionnaire data were available for 201 of the 281 identified breast cancer cases (72%). If the report of breast cancer by a woman (or her next of kin) was specifically contradicted by another source of information , we did not include the report as a case (n = 10), we used the date of death as the date of diagnosis unless timing of diagnosis was otherwise noted on the death certificate, in which case we adjusted the date accordingly.We obtained dates of breast cancer diagnosis from self-report, proxy report, the medical record, cancer registry record, or the death certificate. When diagnosis dates from multiple sources conflicted, we used the earliest date considered most valid . For breast cancer decedents with only death certificate information .We used the NIOSH Personal Computer Life Table Analysis System (PC-LTAS) to determine whether the incidence of breast cancer in the study cohort was higher than that expected from general population rates . We consZ-test with an α of 0.05.Analyses using SEER referent rates produced SIRs and SRRs by categories of cumulative exposure and exposure duration, stratified by age (in 5-year categories), calendar time (in 5-year categories), and, for some analyses, race (white and nonwhite). Follow-up time began in 1970 when the SEER rates became available, or 1 year after employment, whichever was later. Beginning follow-up in 1970 eliminated 24 (8.5%) of the 281 breast cancer cases and 34.3% of the person-time from LTAS analyses only. Follow-up continued until the date of diagnosis, date of death, study end date (31 December 1998), or last known date before loss to follow-up, whichever was earliest. We also conducted lagged analyses to allow for the necessary latency period for solid tumors, with 0-, 10-, and 15-year lags examined. This lag discounts all exposure in the last years before the cutoff date, and sometimes results in a worker having no exposure or being “lagged out.” For exposure–response analyses, LTAS calculated a linear trend in a person-year–weighted regression of directly standardized rates, with statistical significance of each trend determined using a two-tailed We conducted internal exposure–response analyses using Cox regression. We used the SAS PHREG procedure , with agWe limited regression analyses for the full cohort  to exploring exposure metrics and birth cohort effects. We considered several exposure metrics: cumulative exposure (log-transformed and untransformed), duration of exposure (log-transformed and untransformed), peak exposure, and average exposure (cumulative exposure divided by duration of exposure). We analyzed all metrics as continuous and categorical, with lags of 0, 10, and 15 years evaluated. We performed similar analyses for the subset of workers with some questionnaire data (the “questionnaire subcohort”) to see whether risks were similar in the two groups.a priori interest (the “restricted subcohort”). We required members of the restricted subcohort  to have data on parity, age at first live birth, replacement hormone use (ever and age at first use), whether a first-degree female relative had had breast cancer, and ever smoked at least 100 cigarettes. We assessed menopausal status but did not include it as a restricting factor; instead, we imputed the age of 52 for women for whom menopausal status or age at menopause was missing (n = 249).Because of the importance of certain covariates in assessing the relations between PCB exposure and breast cancer risk, we restricted further analyses to workers with complete questionnaire data on covariates of We knew from prior studies  that theWe performed covariate analyses on the restricted cohort, using backward elimination to rule out the least influential of the potential confounders. We also examined key interaction terms, evaluating statistical significance with the chi-square likelihood ratio test.Among the full cohort of 5,752 women, the mean ± SD duration of employment was 7.7 ± 4.5 years. A total of 1,413 (25%) had died by the study end date. We obtained completed questionnaires for 3,952 (69%) of the cohort, 80% from living respondents and 20% from next of kin of deceased cohort members. The primary reason for nonresponse was our inability to determine a correct address (21%). A total of 500 former workers and proxy respondents (8.7%) refused to complete the questionnaire, and 112 (2%) failed to respond after repeated attempts.We found 281 incident breast cancer cases in the full cohort. Although mortality follow-up of the cohort was exhaustive, nonfatal incident cases occurring before initiation of the cancer registries or occurring in states other than Massachusetts, New York, Indiana, Florida, and California could be ascertained only via questionnaire. Approximately 20% of the cohort did not return a questionnaire and were not known to be deceased at end of follow-up; incident cases occurring outside the registry search may have been missed in this group, particularly in the early years of follow-up.Of the full cohort, we identified 147 women by plant records and death certificates as being of races other than white. When we used the questionnaire responses instead, the number of workers (both cases and noncases) identified as being of races other than white rose to 282. This group includes workers identifying as members of specific races other than white , women who identified only as “other,” and women identifying as multiracial. Because the number of workers identifying as other than white was small, we kept this group together for all further analyses and refer to it herein as “nonwhite,” although some women in fact selected a multiracial identity where one of the races was white.a priori interest (restricted cohort), we identified 14 of the 145 cases as nonwhite. Twelve of these were from plant 2. Of the 268 noncase workers identified as nonwhite on the questionnaire, 67 identified as African American, and 201 identified as “other.” Most workers (n = 150) in the last group did not further specify their race/ethnicity; of those who did, 90% (n = 55) identified themselves solely as Cape Verdean or as Cape Verdean and some other ethnicity.Of the 281 cases in the full cohort, we identified only eight as nonwhite from plant or death certificate data. However, among workers with complete questionnaire data on the covariates of n = 5,752) and the questionnaire subcohort . In the entire cohort, on average, cases (n = 281) were born earlier than noncases and were exposed longer; these differences were highly statistically significant. Cases also had higher cumulative exposures. Breast cancer in situ was reported for 8 of the 281 cases (2.8%).n = 201) were born significantly earlier than noncases and were exposed longer. Cases had higher mean cumulative exposure than did noncases, but the difference was not statistically significant. Cases were significantly more likely to have a first-degree relative with breast cancer than were noncases. Average age at first live birth showed a slight but statistically significant difference. There were no statistically significant differences between case and noncases in the average age at menarche, body mass index (BMI), parity, number of children, use of hormone replacement therapy, age at first use of hormone therapy, or in the proportion of nonwhites or ever-smokers.More data were available for the questionnaire subcohort, including demographic and lifestyle characteristics. Again, cases  was more than twice that in plant 1 (0.34 ± 0.71) and was an order of magnitude greater than that in plant 3 (0.078 ± 0.10). The medians reflected a similar pattern . Although all three facilities had nonwhite populations < 4% according to records-based data, we classified > 15% of workers at plant 2 as nonwhite using questionnaire data. The percentages of nonwhite workers for plants 1 and 3 increased only slightly when we used questionnaire data.In the aggregate, white and nonwhite workers in the restricted subcohort differed on a number of demographic and lifestyle factors . Statistn = 257] and using a 10-year lag was 0.81 . SRRs by exposure level showed no trend using lagged or unlagged data. However, the SIRs and SRRs differed somewhat by race. For women identified as white by plant records or death certificates, the SIRs were 0.80  and 0.80  with 0- and 10-year lags, respectively. For women identified as nonwhite by records , the SIRs had nonsignificant elevations of 1.87  and 1.94  for 0- and 10-year lags respectively. SIRs for women with questionnaire data were similar to those for the full cohort in white women  but showed a more modest elevation in nonwhite women  than that seen in the full cohort.SIR and SRR results reflect follow-up time and cases occurring in 1970 or later, because of availability of SEER comparison data. For the full cohort, the unlagged SIR was 0.81 [95% confidence interval (CI), 0.72–0.92; SRR results differed depending on dose cut points and lag (data not shown), particularly for white women, with no consistent patterns observed. In nonwhite women, trends were always positive and sometimes statistically significant, depending on cut points and lag. However, elevations were limited to the highest dose categories, with deficits observed in some intermediate categories.in situ cases.Because the SRR results suggested a difference in risk by race, we evaluated the effects of race in the regression analyses and then conducted further evaluations of potential confounders and effect modifiers in the restricted cohort separately for nonwhite and white women. Among white women, cumulative exposure and duration of exposure had little effect on breast cancer risk. Birth cohort was significant, with an increased risk for the group born during the period 1920–1934 and a larger increase compared with baseline for those born after 1934. Beyond birth cohort, the covariates retained in the model were parity, family history of breast cancer, and self- versus proxy questionnaire completion. Plant did not strongly affect the relation between the exposure metrics and breast cancer risk, and we eliminated this variable from the model. Use of hormone replacement therapy, although technically a confounder, had a lesser effect on risk, and we also eliminated this variable from the final model in the interest of parsimony. Menopausal status was not a confounder in this group.In contrast, among nonwhite women, the effects of increasing exposure were positive and statistically significant. Both cumulative exposure (continuous) and duration of exposure were highly associated with breast cancer risk. Categorical exposure was associated with elevated risk at the two highest levels, but not in the intermediate category. The variables usually considered breast cancer risk factors  had little effect on risk in this group (data not shown). The most important covariates for this cohort were ever-smoking, source of questionnaire data, and birth cohort (limited to those women born after 1934). In univariate analyses, nonwhite cases smoked more than noncases (46 vs. 29 pack-years), in part because they smoked for more years (42.6 vs. 32.9). We did not observe these patterns among white workers. To assess whether smoking might be a proxy for alcohol use, we examined models with smoking but not alcohol, with alcohol but not smoking, and with both alcohol and smoking and found that only smoking had significant explanatory value.p = 0.06).To further explore the effects of smoking in nonwhite workers, we considered current smoking status (at time of death for deceased workers or time of questionnaire completion for living workers) and pack-years. Adding pack-years or current smoking status to a model that included ever smoking had little effect. The interaction between current smoking status and cumulative exposure nearly attained statistical significance , the only covariates with significant explanatory power. Race was not a variable of strong In both white and nonwhite women, risk of breast cancer after PCB exposure was lower for women born before 1920 than for those born after 1934. The pattern of increasing breast cancer incidence risk for later birth cohorts is generally consistent with results reported from the SEER registry  and the The study has several potential limitations. Response bias is a possibility when questionnaires are used. If nonrespondents had lower breast cancer rates than respondents, an artificial excess of breast cancer would be observed in the exposed population versus the nonexposed referent population. To minimize the likelihood of such bias, the study materials referred not to breast cancer but to “health effects.”Of greater concern is that ascertainment of incident breast cancer cases was likely better for the later years of the study, when cancer registry data were available to supplement death records and questionnaire data. Although we identified only 10 breast cancer deaths using death certificates alone, it is likely that nonfatal incident cases were missed in the early years. This would lead to underestimates of the SIR and could affect exposure–response analyses, because early exposures were substantially higher than those in later years. In addition, this could affect observed secular trends. On the other hand, we could not verify some of the cases identified via questionnaire either by medical records or by a cancer registry, so we cannot rule out limited overascertainment.Completion of questionnaires by proxy respondents likely introduced misclassification on some covariates. Proxy respondents often left blanks in response to some questions, so the restricted cohort underrepresents deceased workers. In addition, proxy respondents may also have been less accurate in responses to questions they did answer. When we performed a sensitivity analysis evaluating proxy and self-respondents in the restricted cohort separately, we found that results for each group were quite similar to those for all respondents combined, although risk estimates were slightly higher among self-respondents.Errors in reporting of smoking are potentially problematic, given the results in nonwhite women. Studies have shown that smoking status is reported more accurately by proxies than was number of cigarettes smoked . We usedMisclassification of PCB exposure is another potential limitation. We assessed exposure levels by job title, department, and work process, with limited air sampling data available from the 1970s only. The combination of dermal and inhalation exposures into a single semiquantitative estimate does not account for differences in doses by route of exposure. In addition, the use of total PCBs as the exposure metric may obscure the effects of individual congeners or subgroups.Race, although an important predictor for breast cancer risk after PCB exposure in this study, is not well defined for this cohort. In addition to general concerns about using race in place of more specific information about genetics to explain disease risk , this stCYP1A1 M3 genotypes who smoked > 20 years  or a vehicle for greater but unmeasured PCB exposure through hand-to-mouth ingestion.The lack of any significant overall increase in risk of breast cancer after PCB exposure in this occupational cohort is reassuring and consistent with the results of previous studies of the contributing cohorts . Chance


The relationship between prostate cancer and indicators of nutrition, diet and social factors was evaluated in a case-control study of 166 patients with histologically confirmed prostatic carcinoma and 202 control subjects hospitalized for acute diseases other than malignant, hormonal or urogenital. The relative risk increased with increasing body mass index, men being moderately overweight showing a 2.3 elevated risk, and those grossly overweight an over four-fold higher risk of prostate cancer, when allowance was made for several identified potential confounding factors. Cases also reported more frequent consumption of milk and other dairy products and meat, but no significant difference was noted for vegetable intake. The risk of prostate cancer was unrelated to marital status or indicators of social class based on occupation.


Phytoestrogens are a group of compounds found in plants that structurally resemble the hormone oestradiol, and thus have the potential to act as oestrogen agonists or antagonists. Their potential effects may alter the risk of breast cancer, but only a limited range of phytoestrogens has been examined in prospective cohort studies.o-desmethylangolensin, and equol) using liquid chromatography/mass spectrometry. Data on participants' diet, demographics, anthropometrics, and medical history were collected upon recruitment. All models were adjusted for weight, fat and energy intake, family history of breast cancer, social class, analytical batch, and factors related to oestrogen exposure.Serum and urine samples from 237 incident breast cancer cases and 952 control individuals (aged 45 to 75 years) in the European Prospective into Cancer-Norfolk cohort were analysed for seven phytoestrogens ; among those with oestrogen receptor-positive tumours, the risk of breast cancer was increased with higher levels of urinary equol , P = 0.013).Urinary or serum phytoestrogens were not associated with protection from breast cancer in the European Prospective into Cancer-Norfolk cohort. Breast cancer risk was marginally increased with higher levels of total urinary isoflavones , There was limited evidence of an association between phytoestrogen biomarkers and breast cancer risk in the present study. There was no indication of decreased likelihood of breast cancer with higher levels of phytoestrogen biomarkers, but the observation that some phytoestrogen biomarkers may be associated with greater risk of breast cancer warrants further study with greater statistical power. O-Desmethylangolensin (O-DMA) and equol are metabolites of daidzein, and enterodiol and enterolactone are metabolites of lignans; the parent compounds are metabolised by the gut flora. Both precursors and metabolites can be measured in biological fluids [It has been proposed that phytoestrogens – dietary compounds that structurally resemble the hormone oestrogen – are associated with the development of breast cancer due to their potential to act as oestrogen agonists or antagonists -3. Phytol fluids .in vitro assays than its precursor daidzein, although only about one-third of individuals are reported to be capable of producing equol [The proposed protective mechanisms of phytoestrogens include stimulation of apoptosis, antioxidant activity, inhibition of enzymes associated with carcinogenesis and steroid metabolism, and competitive binding to oestrogen receptors (ERs) -12. Phytng equol ,15.Serum and urinary biomarkers of phytoestrogen exposure serve as an alternative indicator of dietary intake, and are advantageous in that they are not reliant on self-reported dietary information or on food composition databases for phytoestrogen quantification. Furthermore, the use of biomarkers reflects the variation across individuals in terms of gut microflora, which is responsible for the production of equol and O-DMA from its parent compound daidzein, and is influenced by stress, diet, bowel disease, genetics, and antibiotic use among other factors . Studies13C-labelled standards for quantitation, we have previously shown that phytoestrogens in plasma and serum are closely correlated [Using newly developed highly sensitive and rapid methods for the analysis of phytoestrogens by mass spectrometry incorporating triply rrelated  and thatrrelated . In the rrelated . We haveBetween 1993 and 1997, men and women aged 45 to 75 years were recruited from general practitioners' databases for the EPIC-Norfolk study . A healtThe International Classification of Diseases, Injuries, and Causes of Death was applied for the classification of incident cases of invasive breast cancer . Between 1993 and July 2006, there were 237 incident cases of invasive breast cancer identified through the East Anglia Cancer Registry, diagnosed a minimum of 12 months after recruitment in the EPIC-Norfolk cohort.Cases were matched with four control individuals based on age (± 3 years) using the entire EPIC-Norfolk cohort prior to the analysis of serum and urine phytoestrogen data. Serum and urine samples were not available, however, for all of the cases and control individuals; as a result, the number of controls varied from one to four per case. For the present analysis of phytoestrogens, therefore, the data were treated as frequency matched and adjusted for age.All 237 cases and 952 control individuals had completed a food diary and had at least one blood or urine sample available for analysis. The ER status of tumour samples was also obtained from the East Anglia Cancer Registry. ER-positive tumours were detected among 115 cases, ER-negative cases were detected among 27 cases, and ER status information was not available for 95 cases. Seventy-four per cent of cases without ER status were from the first batch; this was probably due to the fact that ER testing was not routinely conducted until 1999. Plasma oestradiol was available for 153 cases and 603 control individuals using methods for estimating plasma oestradiol published previously .13C-labelled internal standards. The average intra-assay coefficient of variation for serum analytes ranged from 2.8% (enterolactone) to 20.0% (glycitein). Limits of detection for serum phytoestrogens ranged from 0.04 ng/ml for daidzein to 0.11 ng/ml for equol, corrected for the concentration of sample [Serum samples, drawn at the health check, were analysed for daidzein, genistein, glycitein, O-DMA, equol, enterodiol and enterolactone. Samples were analysed as described previously . Brieflyf sample . The 284f sample  were desSpot urine samples were analysed for the same phytoestrogens as the serum samples. Three hundred and thirty-three urine samples were analysed using isotope dilution gas chromatography/mass spectrometry, wherein dried extracts were derivatised to their trimethylsilyl derivatives and were designated Batch 1 as before. The remaining 731 samples (Batch 2) were analysed by liquid chromatography/mass spectrometry, in which the dried extracts were redissolved in 40% aqueous methanol and were analysed as above. The limits of detection with gas chromatography/mass spectrometry after correction for dilution of sample ranged from 1.8 ng/ml (enterodiol) to 8.0 ng/ml (enterolactone). For liquid chromatography/mass spectrometry, the limit for detection ranged from 0.08 ng/ml for daidzein to 0.22 ng/ml for equol. The intra-assay coefficient of variation for urinary phytoestrogens ranged from 2.3% (daidzein) to 6.8% (equol). Details of the quality assurance and methodology have been published elsewhere [t test was conducted for continuous variables (log-transformed). To be retained in the log transformation, the value 0.011 was assigned arbitrarily for phytoestrogen levels below the level of reliable quantification, and the value 0.01 was when the compound was not detected at all. Pearson correlation coefficients were calculated to ascertain the relationship between log-transformed urinary and serum measures of each phytoestrogen. The Wilcoxon–Mann–Whitney test was applied to determine whether the probability distribution for serum and urinary phytoestrogens differed between cases and control individuals.For case–control comparisons of potential confounders, a chi-square test was applied to categorical variables and an unpaired Student 2-transformed, so the estimates of risk correspond to a doubling in phytoestrogen level [P value or if the β values for the phytoestrogen in the model were modified by 10% or more when added individually to the model. This approach identified weight (continuous), fat (continuous), energy (continuous), family history of breast cancer , and social class  as covariates. Smoking status , education , fibre intake (continuous), height (continuous), and physical activity level  were found to be nonsignificant covariates and were therefore not included in the final models. Among subjects with oestradiol data available, adjustment for oestradiol (continuous) did not alter the β values for the phytoestrogens under examination and was therefore not included in the final models.Odds ratios for breast cancer risk were calculated using unconditional logistic regression. Summary variables of total lignans (enterodiol and enterolactone) and total isoflavones  were created and analysed in conjunction with individual phytoestrogens. Phytoestrogen values were logen level . All modCovariates were missing for fewer than 5% of the sample unless specified otherwise; subjects with missing covariate data were deleted from the analysis. Analyses were repeated for cases with tumours that were identified as ER-positive and their controls. The small number of participants with ER-negative tumours precluded analysis of this subgroup. To further explore the role of oestrogen, the sample was stratified by menopausal status , and analyses were re-run on the subgroup with oestradiol data available.P values are two-sided; values less than 0.05 were considered significant, with marginal significance noted below 0.1.All analyses were conducted with SAS software . All Demographic and lifestyle characteristics of the 237 breast cancer cases and 952 control individuals are presented in Table P = 0.003. Correlations including values below the level of quantification ranged from 0.27 to 0.58 (P < 0.0001).The median serum and urinary phytoestrogen levels are presented in Table P = 0.022), but was not significant among postmenopausal women , P = 0.372); no change in results was observed for serum phytoestrogens (data not shown). The small number of premenopausal and perimenopausal women in the ER subgroup precluded stratification by menopausal status.None of the serum phytoestrogens were associated with the risk of breast cancer, regardless of ER-positive tumour status. In the full study, an increased risk of breast cancer was observed with higher levels of total urinary isoflavones Table . Among wDespite the biological plausibility of a proposed reduction in breast cancer risk with greater exposure to phytoestrogens -12, the The predominant absence of associations between phytoestrogen biomarkers and breast cancer may be attributed to limitations specific to aspects of the study design. The half-life of daidzein, genistein, and equol has been reported as <12 hours ; the sinIncreased risk with greater levels of urinary equol in the ER-positive subgroup of the present study was among the few significant associations detected; however, this association should be interpreted with caution as it was not also observed in the analysis of serum equol. This result is consistent with the previous study of women in the EPIC-Norfolk cohort, wherein breast cancer risk was positively associated with serum and urinary equol , and witThe multiple comparisons conducted in the present analysis may have created a false-positive finding for urinary equol in the ER-positive group. This methodological concern also applies to the marginally significant associations detected for total urinary lignans in the ER-positive cohort and for total urinary isoflavones in the full study. Additionally, the disparity between urinary and serum results may have been affected by fact that the proportion of participants with undetected values was generally higher for phytoestrogen biomarkers in serum compared with urine. Nonetheless, the evidence suggests that there may be circumstances where phytoestrogen exposure increases breast cancer risk, and thus deserves further studyThe literature regarding a potential interaction between phytoestrogens and ER status in relation to breast cancer is limited and has not included a measure of equol. In a prospective Danish case–control study, plasma enterolactone was not associated with ER-positive breast cancer, but was associated with a decreased risk of ER-negative breast cancer; the authors noted that, with only 80 ER-negative cases, caution is warranted in interpretation of this result . A prospStudies of dietary measures of phytoestrogen in relation to ER-positive cancer risk have been limited by the quality of data available and by the range of phytoestrogens investigated. In a large French prospective study including 1,469 breast cancer cases, ER-positive and progesterone-positive breast cancer risk was inversely related to an estimate for intake of total dietary lignans  . Two stuOnly two prospective studies have examined a wide range of metabolites with the same methods as the present study ,25; all Studies of phytoestrogen intake in western populations have yielded inconsistent conclusions on the nature of the relationship with breast cancer, with evidence of either decreased risk ,40 or noOverall, the absence of strong associations between phytoestrogen biomarkers and breast cancer in the present study could indicate that the level of exposure was too low to yield an effect on cancer development. It has been noted previously that the intake of isoflavones is relatively low in the EPIC-Norfolk cohort . The lowThe majority of the urinary or serum biomarkers for phytoestrogen levels were not associated with breast cancer risk in the EPIC-Norfolk cohort. The limited evidence of an association between urinary equol and ER-positive breast cancer cases warrants further study in subsequent examinations of breast cancer risk and phytoestrogen biomarkers. Overall, the present study does not provide sufficient insight into the nature of the relationship between phytoestrogens and breast cancer risk to warrant modification of phytoestrogen intake among those at risk for breast cancer.o-desmethylangolensin.EPIC = European Prospective into Cancer; ER = oestrogen receptor; O-DMA = The authors declare that they have no competing interests.HW conducted the statistical analyses and wrote the manuscript. GC and GGCK conducted the laboratory analyses of serum and urine phytoestrogens, and contributed to the manuscript. RL participated in the statistical analysis and was responsible for follow-up and identification of cases. K-TK is a principal investigator of the EPIC-Norfolk study and contributed to the manuscript. SB led the study, was a principal investigator in the EPIC-Norfolk study, and drafted the manuscript with HW. All authors read and approved the final manuscript.


Vitamin D receptor (VDR) genotypes may influence breast cancer risk by altering potential anticarcinogenic effects of vitamin D, but epidemiological studies have been inconsistent. Effect modification by serum 25-hydroxyvitamin D (25 [OH]D), the biomarker for vitamin D status in humans, has rarely been examined.VDR-5132 and Cdx2) in the VDR gene, which thus far have not been analyzed with respect to breast cancer risk, on postmenopausal breast cancer risk in a population-based, case-control study including 1,408 patients (cases) and 2,612 control individuals (controls) matched for year of birth. Odds ratios (ORs) for breast cancer adjusted for potential confounders were calculated for genotypes and estimated haplotypes.We assessed the effects of two frequently analyzed polymorphisms (FokI and TaqI) and two potentially functional variants  but not for oestrogen receptor negative tumours . Haplotype analysis revealed the haplotype FtCA , which contains the TaqI t allele, to be associated with a significantly greater breast cancer risk as compared with the most frequent haplotype FTCG . No significant interaction between VDR genotypes or haplotypes and 25(OH)D was observed.No differences in serum 25(OD)D concentrations by VDR polymorphisms on postmenopausal breast cancer risk and possible differential effects of receptor status of the tumour. However, further studies focusing on the influence of polymorphisms and haplotypes on VDR functionality, activity and concentration are needed.Our results support potential effects of  In various observational studies vitamin D intake and serum concentrations of vitamin D metabolites have been associated with decreased risk for developing breast cancer -3. Apart2D). The biologically most active form of vitamin D is 1,25(OH)2D, which mainly exerts its antiproliferative effects by binding to the vitamin D receptor (VDR) and acting in complex as a transcriptional factor for a variety of genes, including those involved in cell differentiation and cell growth D) and to 1,25-dihydroxyvitamin D . The f allele leads to a protein that is three amino acids longer and less effective  study) that was conducted in two regions in Germany: the city of Hamburg and the Rhein-Neckar-Karlsruhe (R-N-K) region. The study was approved by the ethics committees of both the University of Heidelberg and the University of Hamburg, and it was conducted in accordance with the Declaration of Helsinki. All study participants gave informed consent. Patients (cases) were eligible if they had a histologically confirmed primary invasive or Two control individuals (controls) per case were randomly selected from lists of residents provided by population registries and frequency matched by year of birth and study region to the cases. Of the 17,093 controls who met the inclusion criteria, 7,421 (43.4%) participated, 7,521 (44.0%) refused to participate and 2,151 (12.6%) did not respond.Using a standardized questionnaire all participants were interviewed by trained personnel to obtain information on sociodemographic factors and potential breast cancer risk factors.in situ breast cancer cases and 6,657 controls were classified as postmenopausal, of whom 1,559 cases and 3,008 controls were from the R-N-K region. Because of organizational aspects of sample handling, we included only participants from the R-N-K study region.Women who reported their last natural menstrual bleeding at least 12 months before the reference date , a bilateral oophorectomy, or cessation of menses because of radiation or chemotherapy were defined as postmenopausal. Women older than 55 years with unclear menopausal status because of hysterectomy or hormone use were also considered postmenopausal . Premenopausal women and women under age 55 years with unclear menopausal status were excluded from the analysis. In total, 3,464 invasive or Postmenopausal participants from the R-N-K region with DNA samples were included in this analysis, which included 1,408 (90.3%) cases and 2,612 (86.8%) controls.TaqI restriction enzyme . The resultant fragments  were resolved on a 3.2% agarose gel. The FokI, VDR-5132 and Cdx2 polymorphisms were genotyped using Pyrosequencing™ technology  D) in the multivariate model. Statistical haplotype-environment interaction was evaluated using Wald statistics for the respective interaction term in haplo.glm, including interaction terms for all haplotypes simultaneously in the model.Statistical genotype-environment interaction was evaluated with the likelihood ratio test by including a cross-product term of the dichotomous Interaction of genotypes or haplotypes with oestrogen receptor (ER) or progesterone receptor (PR) status of the tumour was assessed by means of a case-only analysis. ER or PR status was used as the dependent variable (outcome) and potential interaction variables (genotypes and haplotypes) as independent variable in the logistic regression model. The Wald statistic for the respective independent variable was used to test for statistical interaction.P ≤ 0.05. All calculations except the haplotype analysis were conducted using the statistical software SAS 9.1 .All tests were two-sided and considered to be statistically significant at In comparison with controls, and consistent with established risk factors for breast cancer, cases were significantly older at menopause, were more likely to have a positive family history of breast cancer, more frequently had a history of benign breast disease, had lower parity, were younger at menarche, were less likely to have breastfed, had a greater number of previous mammograms, and more frequently used hormone therapy Table .Median 25(OH)D concentrations were 44.9 nmol/l and 51.5 nmol/l for cases and controls, respectively. We did not observe significant differences in 25(OH)D concentrations by genotype in any of the four analyzed polymorphisms . The observed allele frequencies in the controls were comparable to those reported in the dbSNP database for Caucasian populations . There was no evidence of linkage disequilibrium between any pair of the four analyzed polymorphism (r2 < 0.01).The genotype distribution in the control group was in HWE for all analyzed polymorphisms (VDR genotypes and breast cancer risk differed by serum 25(OH)D level. No departure from multiplicative interaction was observed when considering serum 25(OH)D level as a continuous variable . Differential effects for the respective polymorphisms were also not found when serum 25(OH)D concentrations were dichotomized at 30 nmol/l, which was defined as the cutt-off point for vitamin D deficiency. Analyses stratified by season did not yield any further significant 25(OH)D-genotype interactions or differences in results between the winter months (October to March) and the summer months (April to September). There was also no significant gene-environment interaction with family history of breast cancer, use of hormone therapy, or smoking status.We did not observe a significant association between genotypes of any of the four analyzed polymorphisms and risk for postmenopausal breast cancer Table . In the P for interaction = 0.04; Table VDR-5132 polymorphism and PR or ER status . However, under the log-additive model, the haplotype FtCA  was significantly associated with breast cancer risk compared with the most common haplotype FTCG . A statistically significant interaction was found for the haplotype FtTA with PR status of the tumour (P for interaction = 0.03), but haplotype frequency was very low (fFtTA = 0.016).Analogous to the genotype analysis, we evaluated statistical interaction of haplotypes with receptor status of the tumour in a case-only analysis. We did not observe an interaction of any haplotype with ER status. For the haplotype FtCA with the significant main effect, the ORs were 1.34 (95% CI = 0.90 to 2.01) and 1.67 (95% CI = 0.94 to 2.99) for ER-positive and ER-negative tumours, respectively D concentration in the subpopulation for whom there were data on 25(OH)D. ORs did not change substantially, but the OR for the risk haplotype FtCA was no longer significant because of lower numbers of cases and controls.VDR gene were associated with postmenopausal breast cancer risk. However, there was effect modification by ER status of the tumour, so that t allele carriage of the TaqI polymorphism was associated with significantly higher breast cancer risk as compared with noncarriage in ER-positive tumours only. Several studies reported no significant association between the TaqI polymorphism and breast cancer risk D and breast cancer risk was not modified by genotype). Our results corroborate the findings in the NHS of an absence of interaction between the FokI SNP and 25(OH)D [VDR might affect 25(OH)D concentration, because the VDR is possibly involved in negative feedback regulation of 1,25(OH)2D synthesis mediated by the 1α-hydroxylase, which is the enzyme that converts 25(OH)D to active 1,25(OH)2D [VDR genotype in our population, which is consistent with results on the relation between FokI and TaqI polymorphisms and vitamin D status in previous smaller studies [We recently reported an inverse association between serum 25(OH)D concentration and postmenopausal breast cancer risk  and ther 25(OH)D . Functio25(OH)2D . However studies ,41.Strengths of our study are the large sample size, the adjustment for all potential breast cancer risk factors, the evaluation of interactions with serum 25(OH)D, and the restriction to postmenopausal women, because various studies so far have not differentiated between premenopausal and postmenopausal women. Genotyping errors can almost completely be excluded, because 10% of the samples were genotyped in duplicate and concordance was 100%. In addition, all analyzed polymorphisms were in HWE and allele frequencies were comparable to those reported in the dbSNP database for Caucasian populations.We selected mainly functional variants in order to assess their effects on breast cancer risk. However, observed associations may have arisen from other unknown functional variants in linkage with the analyzed polymorphisms in our study. In contrast, the haplotype analysis was exploratory because functionality of the analyzed haplotypes is unknown. Therefore, further studies with respect to functionality of haplotypes are necessary.Limitations due to the retrospective case-control design are of less importance when assessing genetic variations, and thus selection bias is unlikely to have biased our results. However, when interpretating the null interactions with serum 25(OH)D, the low response rate in the population controls and measurement of 25(OH)D after diagnosis in the cases may be of concern. Information on 25(OH)D status, diet, or vitamin D related variables such as outdoor activity in nonparticipants was not available. A cancer diagnosis may change dietary or behavioural habits, which may influence 25(OH)D concentrations. Modification of dietary habits after a cancer diagnosis appears to be limited , but casOur findings are not representative for non-Caucasian populations because women in our population are primarily Caucasian and allele frequencies vary widely among populations of different ethnic origin.VDR-5132 C, Cdx2 A) was associated with a significantly higher breast cancer risk as compared with the most frequent haplotype (FTCG). No significant interaction between VDR SNPs or haplotypes and serum 25(OH)D was found. Our results support potential effects of VDR polymorphisms on postmenopausal breast cancer risk. Further epidemiological studies assessing the association of vitamin D and breast cancer risk should take the receptor status of the tumour and other gene variants of oestrogen metabolism into account. In addition, more studies on certain polymorphisms and haplotypes in the VDR, especially functional studies with respect to impact on VDR activity and concentration, are needed.None of the analyzed polymorphisms was associated with risk for breast cancer overall. However, the t allele of the TaqI polymorphism was associated with a significantly increased breast cancer risk in ER-positive tumours only. In a haplotype analysis the haplotype FtCA D = 25-hydroxyvitamin D; CI = confidence interval; ER = oestrogen receptor; HWE = Hardy-Weinberg equilibrium; NHS = Nurses Health Study; OR = odds ratio; PCR = polymerase chain reaction; PR = progesterone receptor; R-N-K = Rhein-Neckar-Karlsruhe; SNP = single nucleotide polymorphism; VDR = vitamin D receptor.1,25(OH)The authors declare that they have no competing interests.SA performed the laboratory analysis and data analysis, and drafted the manuscript. AN contributed to the genotyping analysis and revised the manuscript. JL gave advice on the data analysis and contributed to writing of the manuscript. TS, SK and EJM reviewed the manuscript and participated in designing the study. DF and JC were responsible for the study design, gave advice on the data analysis and revised the manuscript. All authors read and approved the final manuscript.