Of the 28 haplotypes noted, four were encountered in frequencies of 5% or more and accounted for 85% of all haplotypes. Subsequently, ATM haplotyping of high-risk, non-Ashkenazi Jews was performed on 66 women with breast cancer and 76 asymptomatic. One SNP (rs228589) was significantly more prevalent among breast cancer cases compared with controls (P=4 × 10−9), and one discriminative ATM haplotype was significantly more prevalent among breast cancer cases (33.3%) compared with controls (3.8%), (P⩽10−10). There was no significant difference in the SNP and haplotype distribution between asymptomatic high-risk and symptomatic women as a function of disease status. We conclude that a specific ATM SNP and a specific haplotype are associated with increased breast cancer risk in high-risk non-Ashkenazi Jews.While genetic factors clearly play a role in conferring breast cancer risk, the contribution of ATM gene mutations to breast cancer is still unsettled. To shed light on this issue, ATM haplotypes were constructed using eight SNPs spanning the ATM gene region (142 kb) in ethnically diverse non-Ashkenazi Jewish controls ( BRCA1 (MIM# 113705) and BRCA2 (MIM# 600185) mutation carriers , age at diagnosis, age at counseling, and ethnic origin at least three generations back. High risk was assigned based on current accepted criteria . The stuDNA samples were taken from unrelated, healthy, non-Ashkenazi individuals, with no personal or familial history of cancer. These were recruited primarily from among women who escorted the high-risk women but were unrelated to them . All study participants among controls gave their consent for anonymous testing.All the patients in this study are non-Ashkenazi Jews originating from Iraq, the Balkan, and Yemen.Genomic DNA was prepared from anticoagulated, venous blood samples using the PUREGene DNA isolation kit  using the manufacturer's recommended protocol.BRCA1  and BRCA2 (6174delT) mutations reported in Jewish individuals. Detection of these mutations was carried out by modified restriction enzyme digest assay, which distinguishes the mutant from the wild-type allele, using primer sequences, cycling profiles, PCR conditions, and gel electrophoresis as previously described (None of the (high risk and control) participants of this study carried any of the four recurring ATM genotyping was performed by PCR amplification of eight SNPs throughout the gene. The SNPs were chosen from three databases: www.ensembl.orgwww.genome.ucsc.edu and www.ncbi.nlm.nih.gov. The SNPs genotyped were ss4328153 (now rs3092993), rs228589, rs600931, rs664677, rs227069, rs664982, rs652541, and rs170548 , and 1 pmol each of forward and reverse PCR primer. After an initial denaturation at 94°C for 5 min, 45 cycles of 94°C for 20 s, annealing at 56°C for 30 s, and extension at 72°C for 1 min were carried out with a final extension period of 3 min. Primer sequences, designed using the software SpectroDESIGNER , are shown in PCR amplification was performed in 384-well microplates , in a total volume of 5 PCR amplification was performed in multiplex reactions as follows:Multiplex 1: SNPrs228589+SNPrs3092993+SNPrs170548.Multiplex 2: SNPrs652541+SNPrs227069+SNPrs664982.Multiplex 3: SNPrs664677+SNPrs600931.Following PCR, SNP genotyping proceeded as previously described . The statistic CCPmax is defined to be maxj∑iPi,j. The P-value is calculated by a permutation test as mentioned above, with the difference of randomly permuting the labels within each population independently, and using the statistic CCPmax instead of CCmax. This statistic avoids the bias in the P-value that might occur due to the mixture of different populations.Since there are three different population groups in the study, originating from Iraq, the Balkan and Yemen, the score for each marker was calculated as follows: Let The permutation test can be readily generalised to handle association between haplotypes and the disease, for example, by adding block haplotypes as artificial loci with states corresponding to common haplotypes. Similarly, one can add loci interactions as artificial loci, whose states are the allele combinations.Estimates of odds ratio (OR), relative risk (RR), and 95% confidence interval (CI) were calculated for the discriminative SNPs and haplotypes  and 3B.Association test for each SNP separately and for the haplotype composed from all the eight SNPs.Association test of all possible pair-wise interactions of SNPs.We repeated the above procedure to perform two different tests:n=34 or 51.5%), Balkan (n=16 or 24.25%), or Yemenite origin (n=16 or 24.25%). Additional 76 (53.5%) women were asymptomatic, high-risk women. The age range at counseling for the high-risk asymptomatic group was 28–67 years (mean 50.3±10.5 years), and their ethnic distribution was as follows: 34 of Iraqi (44.7%), 26 of Balkan (34.2%), and 16 of Yemenite origin (21%). The ethnicity of the population-matched healthy controls (n=118) was 55 (46.61%) Iraqis, 29 (24.57%) of Balkan origin, and 34 (28.81%) Yemenites. The age range at counseling for the control group was 33–69 years (mean 53.4±8.9 years).In total, 260 unrelated women of non-Ashkenazi Jewish origin were genotyped in this study. Of these, 142 were evaluated and considered at high risk for developing breast cancer based on their family history. Of the high-risk individuals, 66 (46.5%) were breast cancer cases (mean age at diagnosis was 48.3±9.7 years) and their ethnic origin was Iraqi  and in BRCA2 (6174delT).All participants were genetically prescreened and shown not carry any of the four common Jewish mutations in Phasing the genotypes of the eight SNPs at the ATM locus yielded together 28 different haplotypes in one block of high LD . Four haP=4 × 10−9). In this SNP, nucleotide T appears in 3.8% of the controls and in 33.3% of the cases (We first compared the breast cancer patients to the control population. SNP 2 (rs228589) is the most associated with the disease, with score of 61.61 (after multiple testing correction: he cases . Odds raAn association test was performed for each SNP separately and for the complete 8-SNP haplotype. The Pearson scores of association are presented in P⩽10−10). We therefore examined the short genotypes consisting of SNP 1 and SNP 2. These genotypes form three common haplotypes, coined A, B and C  and 2 (rs228589) (both intronic SNPs)  (P=7 × 10−9). When testing all pairwise interactions of SNPs, the most associated pair is SNP 2 (rs228589) and SNP 3 (rs600931) (P⩽10−9)  . One rar(P⩽10−9)  confers vs cases yields no significant result (P=0.35).Testing association of the individual SNPs and of all the SNP pairs on the group of high-risk women In this study, several ATM SNPs were seemingly associated with breast cancer risk in Jewish non-Ashkenazi women at high-risk for breast cancer. These results further establish ATM as a contributor to breast cancer susceptibility in high-risk populations.Increased breast cancer risk in ATM heterozygote mutation carriers has been previously reported in studies that either inferred obligate carriership . InteresTwo ATM germline alterations (Ala2524Pro and 6903insA) reported in A-T families have been shown to segregate with breast cancer in these families  was noted by testing each SNP separately, and correcting for multiple hypotheses using permutation tests. Given the intronic position of the two SNPs most tightly associated with breast cancer risk and phenotype, it is unlikely that these SNPs in and by themselves are disease associated. Rather, in all likelihood they are in linkage disequilibrium with a pathogenic ATM mutation.The current study is the first to report ATM SNP and haplotype in a population of high-risk non-Ashkenazi Jewish women. Unlike the lack of a discriminating ATM haplotype among average risk Ashkenazi Jewish breast cancer women (P-value takes into account the three different subpopulations and corrects for multiple testing. Hence, the strong association noted between the ATM genotype and the high-risk phenotype seems real, and cannot be accounted for as an artifact caused by analysis of related individuals.It is important to emphasise that only one patient was analyzed per high-risk family, so that patients in the high-risk group are not more genetically related to each other than in the control group. Additionally, our statistical method for computing the The limitations of the study should be pointed out. This was a relatively small study that analyzed a highly selected population, and includes only non-Ashkenazi Jewish women who were recruited through high-risk clinic in a single medical center in Israel. Thus, the applicability of the results to average-risk population or even high risk, ethnically diverse populations, needs to be established.In conclusion, the present study suggests that a specific ATM SNP seemingly contributes to breast cancer predisposition in Jewish non-Ashkenazi high-risk women in Israel.


The elevated risk was observed for all types of meat and fish intake. Intake of eggs and milk was not related to risk. Cooking methods and doneness levels for meat and fish were not associated with risk, nor did they modify the association with meat and fish consumption. Our study suggests that animal food consumption may play an important role in the aetiology of endometrial cancer, but cooking methods have minimal influence on risk among Chinese women.We evaluated animal food intake and cooking methods in relation to endometrial cancer risk in a population-based case–control study in Shanghai, China. A validated food frequency questionnaire was used to collect the usual dietary habits of 1204 cases and 1212 controls aged 30–69 years between 1997 and 2003. Statistical analyses were based on an unconditional logistic regression model adjusting for potential confounders. High intake of meat and fish was associated with an increased risk of endometrial cancer, with adjusted odds ratios for the highest  In vivo and in vitro studies have suggested that meats cooked at a high temperature generate heterocyclic amines (HCAs) and other potent mutagens and carcinogens, which can induce tumours in multiple tissue sites in many animal species , 66 who died before interview (4.5%), 23 (1.6%) who could not be located, 12 who were out of town during the period of interview (0.8%), and 14 who could not be interviewed for other miscellaneous reasons (1.0%). The median interval between diagnosis and interview was 5.6 months.n=340, 20.9%), being out of town during the period of interview , severe illness , and other miscellaneous reasons .Controls were randomly selected from the Shanghai Resident Registry and frequency matched to cases according to age distribution of endometrial cancer cases in 1996. After exclusion of 63 women who reported a hysterectomy, 1212 out of 1629 eligible controls (74.4%) consented and were interviewed for the study. Reasons for non-participation among controls were refusal  used in the study has been validated by two FFQ surveys and one 24-h diet recall survey  and their 95% confidence intervals (CIs). All of the ORs were adjusted for age (as a continuous variable), menopausal status , diagnosis of diabetes (ever/never), alcohol consumption (ever/never), body mass index (BMI) (as a continuous variable), physical activity in metabolic equivalent tasks (as a continuous variable), and total energy intake (as a continuous variable). Other variables, including dietary intake of total vegetables and soy foods, did not appreciably alter the risk estimates and were not adjusted for in our analyses. Tests for trend across quartiles were performed in a logistic regression model by assigning a numerical score (1–4) to each category of categorised variables. Stratified analyses were conducted to evaluate potential modifying effects.Consumption of common animal foods was categorised into four groups based on the quartile distribution among controls. Uncommon animal foods, such as organ meat, eel, shellfish, and milk were grouped into two levels, never and ever. The Wilcoxon rank sum test was used for comparisons of the median difference between cases and controls and the Presented in The associations between animal food intake and endometrial cancer risk are presented in Neither cooking method nor doneness level of meat and fish were associated with endometrial cancer risk, nor did they modify the association of meat and fish intake with disease risk . AdditioIn vitro and in vivo studies have found that omega-3 fatty acids, which fish, particularly marine fish, are rich in, inhibit the proliferation and progression of hormone-related cancer  is lower than that in cases (82.8%), raising a concern about selective participation bias. Although neither the potential study participants nor the interviewers knew our study hypotheses, we cannot exclude the possibility that study participation would be influenced by general health knowledge including healthy dietary habits. However, we found that the controls of the current study had patterns of meat and fish intake similar to the controls of the Shanghai Breast Cancer Study which was conducted in the same population 2–3 years before the current study which had a much higher response rate (91.1% for cases and 90.3% for controls) . In ChinIn summary, this large population-based case–control study of endometrial cancer, conducted in a population that has a much lower total meat consumption compared to people in Western countries, found that diets high in animal foods increase the risk of endometrial cancer.


Routinely collected data for New South Wales were used to analyse cancer mortality in migrants born in East or Southeast Asia according to duration of residence in Australia. A case-control approach compared deaths from cancer at particular sites with deaths from all other cancers, adjusting for age, sex and calendar period. Compared with the Australian-born, these Asian migrants had a 30-fold higher risk of dying from nasopharyngeal cancer in the first 2 decades of residence, falling to ninefold after 30 years, and for deaths from liver cancer, a 12-fold risk in the first 2 decades, falling to threefold after 30 years. The initial lower risk from colorectal, breast or prostate cancers later converged towards the Australian-born level, the change being apparent in the third decade after migration. The relative risk of dying from lung cancer among these Asian migrants was above unity for each category of duration of stay for women, but at or below unity for men, with no trend in risk over time. An environmental or lifestyle influence for nasopharyngeal and liver cancers is suggested as well as for cancers of colon/rectum, breast and prostate. © 1999 Cancer Research Campaign


A hospital-based case-control study of ovarian cancer was conducted in London and Oxford between October 1978 and February 1983. Menstrual characteristics, reproductive and contraceptive history and history of exposure to various environmental factors were compared between 235 women with histologically diagnosed epithelial ovarian cancer and 451 controls. High gravidity, hysterectomy, female sterilisation and oral contraceptive use were associated with a reduced risk of ovarian cancer. Infertility and late age at menopause were associated with an increase in risk. While these factors were related, they were each found to be independently associated with ovarian cancer risk after adjusting for the effect of the other factors.


A recent study showed an association between a single base substitution, T-->C, in the promotor region of the CYP17 gene, the risk of breast cancer and age at menarche in Asian, African-American and Latino women from California and Hawaii. The C allele was associated with increased risk of breast cancer, significantly so for patients presenting with advanced disease, whereas the TT genotype was associated with later age at menarche in control subjects. We attempted to confirm these findings in a large case-control study in East Anglia, England (835 cases and 591 control subjects). We found no evidence of an increased risk of breast cancer  or advanced breast cancer  in C allele carriers, nor any association between age at menarche and genotype. We conclude that these alleles do not significantly alter breast cancer risk in the English population.


Ptrend=0.003 for oestradiol; 3.24 (1.87–5.58), Ptrend<0.001 for oestrone; 2.37 (1.39–4.04), Ptrend=0.002 for testosterone; 2.07 (1.28–3.33), Ptrend<0.001 for androstenedione; 1.74 (1.05–2.89), Ptrend<0.001 for dehydroepiandrosterone sulphate (DHEAS); and 0.51 (0.31–0.82), Ptrend<0.001 for SHBG. Analyses limited to the 191 cases who had donated blood five to 12.7 years prior to diagnosis showed results in the same direction as overall analyses, although the tests for trend did not reach statistical significance for DHEAS and SHBG. The rates of change per year in hormone and SHBG levels, calculated for 95 cases and their matched controls who had given a second blood donation within 5 years of diagnosis, were of small magnitude and overall not different in cases and controls. The association of androgens with risk did not persist after adjustment for oestrone , the oestrogen most strongly associated with risk in our study. Our results support the hypothesis that the associations of circulating oestrogens with breast cancer risk are more likely due to an effect of circulating hormones on the development of cancer than to elevations induced by the tumour. They also suggest that the contribution of androgens to risk is largely through their role as substrates for oestrogen production.We assessed the association of sex hormone levels with breast cancer risk in a case–control study nested within the cohort of 7054 New York University (NYU) Women's Health Study participants who were postmenopausal at entry. The study includes 297 cases diagnosed between 6 months and 12.7 years after enrollment and 563 controls. Multivariate odds ratios (ORs) ) for breast cancer for the highest quintile of each hormone and sex-hormone binding globulin (SHBG) relative to the lowest were as follows: 2.49 (1.47–4.21),  Nine prospective studies have now reported on the association between endogenous sex hormone levels in postmenopausal women and subsequent breast cancer risk . RemainiBetween 1985 and 1991, the NYU Women's Health Study enrolled 14 275 healthy women aged 34–65 years at the Guttman Breast Diagnostic Institute, a breast cancer screening centre in New York City . Women wBreast cancer cases were identified through active follow-up of the cohort by mailed questionnaires approximately every 2–4 years and telephone interviews for nonrespondents, as well as record linkage with state cancer registries in New York, New Jersey, and with the US National Death Index. A capture–recapture analysis estimated the ascertainment rate in our cohort to be 95% , date of enrollment (±3 months), and number and dates (±6 months) of subsequent blood donations, if any. Menopausal status was confirmed by measuring follicle-stimulating hormone (FSH) in all women for whom the lagtime between last menstrual period and blood donation was less than 2 years and all women who were less than 60 years old at entry and reported having had a hysterectomy without complete bilateral oophorectomy and women with FSH levels ⩽12.75 mIU ml−1), 6.7 and 14.4% for oestrone (at 74 pmol l−1), and 7.8 and 13.5% for androstenedione (at 1.4 nmol l−1), 8.7 and 15.8% for testosterone (at 1.4 nmol l−1), 5.4 and 14.7% for DHEAS (at 1.62 μmol l−1), and 5.6 and 13.5% for SHBG (at 40 nmol l−1).All assays were conducted in the Hormones and Cancer Group at the International Agency for Research on Cancer in Lyon, France. Assays were selected based on the results of a validity study , serum measurements were categorised into quintiles, using the frequency distribution of the cases and the controls combined. The matched set data were analysed using conditional logistic regression . Odds rarmations . All P-vTo explore whether the presence of a growing cancer results in an increase in circulating levels of sex hormones, we calculated the rate of change per year in hormone and SHBG levels in the subset of subjects for whom two blood donations were available and the second blood donation was within 5 years of the index date. We compared values in cases and their matched controls using a mixed-effects regression model. These analyses were controlled for age through the matching. We also controlled for baseline level of hormone (or SHBG), rate of change per year in body mass index, and time since menopause.By 1 March 1998, the start date of the latest round of follow-up, 306 participants postmenopausal at enrollment were first diagnosed with invasive adenocarcinoma of the breast 6 months or more after entry into the study. Nine cases (3%) were excluded for the following reasons: postmenopausal status not confirmed by FSH analysis (two cases), lack of serum (two cases), both selected controls developed cancer (breast or other) and their serum was reserved for analyses in which they were the index cases (five cases). The remaining 297 cases are included in the present analysis. Pathology reports were obtained for 232 cases (78%), and 51 additional cases (16%) were confirmed by the Tumour Registries. Among the 594 controls initially selected, 31 (5%) were excluded for the following reasons: postmenopausal status not confirmed by FSH analysis (four controls), participant had been selected as a control for a previous case (seven controls), participant developed breast (six controls) or another cancer (14 controls), and her serum was reserved for analyses in which she was the index case.vs 23%, P=0.09), a higher frequency of history of breast biopsy , a higher median weight , and higher median body mass index .P<0.001).P=0.09) became significant after adjusting for baseline androstenedione level, rate of change in body mass index and time since menopause (P=0.01).A second blood sample collected within 5 years of the index date was available for cases and controls from 95 matched sets. The mean duration between first and second blood donations was 31 months , and the mean duration between second blood donation and diagnosis was 28 months . In 1995, we reported a positive association between endogenous oestrogens and breast cancer risk based on the first 130 cases observed among the postmenopausal participants in the NYU Women's Health Study .Our initial results on total oestradiol and oestrone levels  because α with greater affinity than oestrone =1.87–5.58; model b) for women in the top quintile, appeared more strongly associated with risk than oestradiol . This result is consistent with the results of the pooled analysis of prospective studies where the OR for women in the top quintile of oestradiol was 2.00 (95% CI=1.47–2.71) and for women in the top quintiles of oestradiol and oestrone were 2.19 (95% CI=1.48–3.22), respectively . As oeststrogens . The strstrogens  have preWhereas oestrogens are known to directly stimulate breast cell proliferation, it is not clear whether the role of androgens is only as precursors of oestrogens, or whether they have a direct role in breast cancer development through conversion into oestrogens in the breast itself or by direct stimulation of the growth and division of breast cells. Multivariate analysis, which is mostly used to control for confounding, may also be used to assess underlying mechanisms : If an avs 1.26 in our study), 1.15 for DHEAS (vs 1.16), and 1.32 for testosterone (vs 1.17). Oestrone-adjusted ORs were not presented in that study. Whether the contribution of androgens to breast cancer risk is direct or indirect, it would be of interest to identify the source of elevated levels of androgens. The increased DHEAS serum concentrations in women who develop breast cancer suggest increased adrenal androgen secretion. In a prospective study including 53 postmenopausal cases, β-hydroxyandrostenedione ratio, which is depressed when the adrenals are the primary source of androstenedione but elevated when the ovaries are the primary source. They concluded that both the adrenals and the ovaries appear to contribute to elevated androstenedione levels in postmenopausal women. Further research on the factors that contribute to elevated androgen levels is warranted.The oestradiol-adjusted ORs associated with a doubling of androgen levels that we observed were very similar to those observed in the pooled analysis of prospective studies , that isThe weak association of breast cancer risk with DHEAS that we observed in our initial analysis . As poinAdjusting for body mass index resulted in an attenuation of the ORs associated with oestrogen levels. These results were expected in the light of the role of adipose tissue in producing oestrogens in postmenopausal women. Spearman's correlation coefficients of body mass index with oestradiol and oestrone were 0.39 and 0.34, respectively. Although smaller, an attenuation of the ORs associated with androgen levels was also observed. A positive association of obesity with increased levels of testosterone has been reported . ObesityTo further investigate whether elevated levels of sex hormones are involved in the induction of breast cancer and not simply a byproduct of the tumour, we conducted an analysis limited to the 191 cases who had donated blood 5 or more years prior to diagnosis. This analysis showed results in the same directions as overall analyses, although the tests for trend did not reach statistical significance for DHEAS and SHBG. Our study is the first to assess the association of breast cancer risk with circulating levels of sex hormones measured well before diagnosis in a fairly large group of cases. The results suggest that the observed associations are more likely due to an effect of circulating hormones on the development of clinical cancer than to an increase in circulating hormone levels induced by the tumour.The availability of a second measurement within 5 years of the index date for cases and controls from 95 matched sets allowed us to examine changes in the levels of hormones and SHBG prior to the index date. If the elevated levels of hormones observed in case subjects were due to the presence of tumours, then these levels would be expected to increase at a faster rate in cases (as they approach diagnosis) than in controls. The mean rate of change of androstenedione indicated a slight increase in serum levels in cases but not in controls. However, for all other hormones and SHBG, changes were of negligible amplitude and not significantly different in cases and controls. These results suggest that the presence of the growing tumour does not have a major effect on circulating levels of sex hormones or SHBG.For completeness and to maximise statistical power, we included in the present analysis all the case subjects in our previous report if diagnosed after 6 months in the study, namely, 82 cases  and their controls included in our initial report were reassayed and included in this report. Excluding these subjects did not materially affect the results (data not shown).In conclusion, our results show that the associations between oestrogen and androgen levels and breast cancer risk are present 5 or more years prior to diagnosis and therefore more likely represent an effect of circulating hormones than of the tumour. This is a key finding towards establishing that sex hormones are causally related to breast cancer. Our results also suggest that the contribution of androgens to breast cancer risk is largely through their role as substrates for oestrogen production.


Until the end of follow-up (1 January 1996), 397 postmenopausal breast cancer cases were identified and a subcohort of 424 women was then taken from all eligible women. Women using hormones were excluded, leaving 364 breast cancer cases and 382 women in the subcohort for the analyses. Concentrations of oestrone, oestradiol, testosterone, 5α-androstane-3α, 17β-diol and creatinine were measured in first morning urine samples, which had been stored since enrolment at −20°C. A Cox proportional Hazards model was used, with Barlow's adjustment for case–cohort sampling, to estimate breast cancer risk in quartiles of each of the, creatinine corrected, hormone levels, the lowest quartile being the reference group. Women with higher levels of all four of the hormones were at increased risk for postmenopausal breast cancer . In conclusion, women with higher excretion levels of both oestrogens and androgens have an increased risk of breast cancer.To assess the relation between urinary endogenous sex steroid levels and the risk of postmenopausal breast cancer, a nested case–cohort study was conducted within a large cohort (the DOM cohort) in the Netherlands ( Established risk factors for breast cancer, such as early age at menarche, late age at menopause, late age at first full-term pregnancy or nulliparity, indicate that endogenous sex hormones play an important role in breast cancer aetiology .For almost 30 years now, several case–control and cohort studies addressed this topic. Recently, a pooled analysis of nine prospective studies, including 663 breast cancer cases and 1765 controls, showed that postmenopausal women in the highest quintiles of oestrone-, oestradiol- and testosterone concentrations in blood had a statistically significant two-fold increased risk of breast cancer . Howeverα-androstane-3α, 17β-diol) and the risk of postmenopausal breast cancer, we conducted a nested case–cohort study in a large population-based prospective cohort study in the Netherlands.To further evaluate the relation between urinary endogenous sex steroids  between 1975 and 1986 (n=9349). All women were followed until 1st January 1996 for the occurrence of breast cancer through their general practitioners and, from 1986 onwards, through linkage with the regional cancer registry. A nested case–cohort study was designed to study the relation between endogenous sex steroids and the risk of postmenopausal breast cancer. Until the end of follow-up, 397 breast cancer cases were identified and a random sample of 424 women (the subcohort) was taken from all eligible women. A urine sample could not be found for three cases and three women from the subcohort. Women using hormone replacement therapy or oral contraceptives at the time of urine sampling were excluded (30 cases and 38 women from the subcohort). Urine samples were sent to the International Agency for Research on Cancer (IARC), Lyon, France, for hormonal measurements. One sample was lost during the analyses and was excluded, leaving 364 cases and a subcohort of 382 women for the analyses.Women who were naturally postmenopausal at recruitment  and who had no history of breast cancer were eligible for the present study  were measured by radioimmunoassay (RIA) after enzymatic hydrolysis, solid phase extraction and high-performance liquid chromatography (HPLC) purification of the urine samples. Results were expressed in nanogram analyte per litre. The method used in this study has been described in detail elsewhere  and between the second and the third collection was 4.0 years (1.3–7.7 years). Intraclass correlation coefficients for sample replicates were above 0.93 for all four hormone metabolites, indicating good reproducibility from a laboratory error perspective. The intraclass correlation coefficients over time were 0.59 for E1, 0.62 for E2, 0.85 for TST and 0.55 for 3αD, indicating reasonable stability of women's urinary hormone levels over several years , oestradiol (E2), testosterone (TST) and 5Creatinine was measured in each sample by kinetic Jaffé reaction .To correct for between-subject variability in urine dilution, in each sample the hormone concentration (expressed in ng/l) was divided by the creatinine concentration (expressed in mg/l). All analyses included creatinine-adjusted hormone levels.Concentrations of all four hormone-metabolites were logarithmically transformed to produce approximately normal distributions. Geometric mean values and their 95% confidence intervals were then calculated in cases and in the subcohort.αD, based on the distribution of the creatinine-corrected hormone levels in the subcohort. Women with hormone levels below the detection limit  were included in the lowest quartile. Incidence rate ratios (IRR) were calculated in each quartile compared to the first, using a Cox Proportional Hazards Model with Barlow's weighting method  oestrogen- and androgen levels.Women were classified into quartiles of concentrations of E1, E2, TST and 3To evaluate the effect of latent disease at urine collection on breast cancer risk, we studied the association between levels of hormone concentrations and the risk for breast cancer for those women who were free of breast cancer for at least 2 years after urine sample collection. To evaluate the influence of the time interval between urine collection and the diagnosis, analyses were done separately for women who developed breast cancer within the first 9 years of follow-up and after 10 years. Linear trends were tested by including quartile scores 1, 2, 3 and 4 as a continuous variable in the model.Cox Proportional Hazards models were performed using the SAS macro described by The median follow-up in the subcohort was 19 years (range 0–21 years). Compared to the subcohort, women with breast cancer more often reported a positive family history of breast cancer and were slightly heavier (αD adjusted for family history of breast cancer. Compared to the lowest quartiles of oestrogens (E1 and E2), women with higher levels of these metabolites were found to be at higher risk of developing breast cancer . Tests for trend were statistically significant for both oestrogens.αD had an increased risk of breast cancer , with statistically significant tests for trend. Analysis of hormone levels unadjusted for creatinine, showed similar results, but relative risks were slightly smaller and not all statistically significant anymore.Also, women with higher levels of both testosterone and 3Additional adjustment for BMI did not change the risks associated with any of the hormone metabolites . ExcludiThe present prospective study shows that postmenopausal women with higher excretion levels of oestrogens as well as androgens have an increased risk of breast cancer.n=364) and follow-up was long (median time to diagnosis=9.8 years). The largest prospective study so far published included 156 cases and a median time to diagnosis of 2.6 years  in most cases  and urinary androstanediol  (in vitro studies (For elevated levels of the nonaromatisable androgen 3Since adipose women have a higher conversion from androgens into oestrogens in the adipose tissue, it has been hypothesised that BMI might actually be in the aetiological pathway, rather than act as a confounder. This may be the case for the reproductive risk factors as well . AdjustmIn conclusion, this study, the largest prospective study on endogenous hormones and breast cancer risk so far, shows that women with higher levels of oestrogens and androgens after menopause have an increased risk for breast cancer compared to women with lower concentrations.


P-value for heterogeneity=0.0001); and the relative risks per 5-year increase in age at first birth were 1.10 (1.07–1.12), 1.23 (1.17–1.29), and 1.13 (1.03–1.23), respectively . Increasing parity reduced the risk of each tumour type, except medullary cancers, but the reduction in risk was greater for mucinous cancers than for any other subtype considered (P<0.05 for comparison with each other subtype in turn). The effect of menopause did not vary significantly by tumour histology. Meta-analysis of published results on the effects of age at menarche and age at first birth on ductal and lobular cancers were in keeping with our findings.Little is known about how reproductive factors affect the risk of breast cancers of different histology. In an analysis of prospective data on 1.2 million middle-aged UK women, we used proportional hazards models to estimate the relative risks of six histological types in relation to menarche, childbearing and menopause. During 8.7 million person-years of follow-up, 17 923 ductal, 3332 lobular, 1062 tubular, 944 mixed ductal lobular, 330 mucinous and 117 medullary cancers were diagnosed. The effect of both age at menarche and age at first birth was greatest for lobular tumours; relative risks per 5-year increase in age at menarche for ductal, lobular, and tubular cancers were 0.93 (0.87–0.99), 0.65 (0.56–0.76), and 0.75 (0.57–0.98), respectively ( Despite the fact that reproductive factors are among the most established risk factors for breast cancer, little is known about whether their effects differ by tumour histology. Few studies of breast cancer in relation to reproductive factors have had sufficient power to estimate reliably the risk of specific subtypes according to detailed reproductive history. We report here on the relationship between age at menarche, parity, age at first birth and age at menopause, and the risk of six histological subtypes of breast cancer in a large prospective study of women in the UK, and present meta-analyses of published results.http://www.millionwomenstudy.org. All participants were flagged on the National Health Service (NHS) Central Registers so that cancer registrations and deaths are routinely notified to the study investigators. These registers provide information on the date of each such event and the cancer site and morphology codes using the 10th revision of the International Classification of Diseases (ICD10)  . All parThe relationship between the risk of incident invasive breast cancer and various reproductive factors was examined separately for six histological subtypes: ductal (ICD10-O code 8500/3), lobular (ICD10-O code 8520/3), mixed ductal lobular (ICD10-O code 8522/3), tubular (ICD10-O code 8211/3) medullary (ICD10-O code 8510/3), and mucinous (ICD10-O code 8480/3). The variables for these analyses were derived from the information provided at baseline, except for those related to menopause, and to hormone-replacement therapy (HRT) use, which were updated from the resurvey, wherever possible. At each point of contact, a woman's menopausal status and HRT use was defined according to the criteria outlined previously . For all2 , family history of breast cancer , average alcohol consumption , and where appropriate, for age at menarche , parity , age at first birth  and time since menopause . Time since menopause and HRT use were treated as categorical time-dependent variables. Women with missing values for a given adjustment variable were assigned to a separate category for that variable. To assess the effect of including women with missing information for some adjustment factors, analyses were repeated for those with known values for all adjustment variables.Proportional hazards models were used to estimate the relative risk (RR) of developing specific histological types of breast cancer according to age at menarche, parity, age at first birth, menopausal status, and age at menopause with attained age as the underlying time variable. All analyses were routinely stratified by broad geographical region (10 regions corresponding to the 10 areas covered by the cancer registries providing data), and adjusted for quintiles of socioeconomic group based on deprivation index  were followed up for incident breast cancer over 8.7 million person-years. During this period, 27 397 were registered with breast cancer with a median age at diagnosis of 60 years (inter-quartile range: 56–64). The three commonest tumour types were ductal , lobular (12%) and tubular (4%) . Mucinouχ22=17.8; P=0.0001; The RR of each of the three commonest types of breast cancer, that is ductal, lobular and tubular tumours, decreased with increasing age at menarche. However, the magnitude of the trend in RR per 5-year increase in age at menarche varied considerably by type . The decrease in the risk of mixed ductal lobular cancers  was similar to that for ductal  and lobular cancer . There was, however, a greater decrease in risk with increasing parity for mucinous cancer , than for the other types (P<0.05 for comparison with each other subtype in turn). In contrast to the decrease in risk with increasing parity seen for five of the tumour subtypes, there was a significant increase in the risk of medullary cancer  . The increase in risk associated with each 5-year delay in age at first birth was greatest for lobular cancer .Every histological subtype of breast cancer examined, except medullary tumours, showed an increased risk with increasing age at first birth among parous women. The RRs by age at first birth for ductal, lobular and tubular cancer, are summarised in χ22=5.1; P=0.08). Nor was there any significant difference, among postmenopausal never-users of HRT, in the trend in risk associated with a 5-year increase in age at menopause across the three commonest subtypes , although once again the trend was least marked for ductal cancers. There were too few cases of mucinous and medullary breast cancers among pre- and perimenopausal women to reliably assess their relationship with menopausal status but among postmenopausal never users of HRT, increasing age at menopause was associated with a non-significant trend of decreasing risk for mucinous cancer , and of increasing risk for medullary cancer .Around 50% (601 568) of women included in the main analyses contributed to the analyses of the effect of menopause among never-users of HRT. Among such women, the relative risk in postmenopausal compared with premenopausal or perimenopausal women was consistently lower than unity for each of the three main types of breast cancer , AlthougThe relationships between the various reproductive factors and the three most common histological types of breast cancer considered here are summarised in A meta-analysis of previous studies that have assessed the effect of age at menarche and/or age at first birth on the risk of both ductal and lobular breast cancer, together with the findings presented here, is shown in In this large prospective study including 27 397 women with incident invasive breast cancer, we have shown that certain aspects of reproductive history are associated with different risks of developing different histological types of breast cancer. In particular, the well-known effects of age at menarche and of age at first birth vary significantly across tumour type, with the greatest effects seen for lobular breast cancer. Increasing parity was associated with reductions in risk for ductal, lobular, tubular and mucinous cancers, with a substantially greater effect for mucinous cancers than for the other tumour types. By contrast, the risk of medullary cancer increased with increasing parity. For most of the reproductive factors considered, the relative risks for mixed ductal lobular cancer were intermediate between those found for ductal and lobular cancer.Several studies , our findings suggest that several indices of reproductive history may have a qualitatively different relationship with risk compared with other breast cancer subtypes. For example, medullary cancer risk increased with increasing age at menarche and with increasing parity but decreased with increasing age at first birth – the opposite of what is found for the more common subtypes. The most notable difference in the effect of parity on medullary compared with all other subtypes examined here was statistically significant and as far as we are aware, no previous study has reported such a difference.The mechanism by which reproductive factors influence the development of breast cancer is not yet well understood, although it is widely believed that the protective effects of a birth may be due to the hormonal changes that occur during pregnancy and lactation . HRT resThe Million Women Study is well placed to investigate the relationships between reproductive history and specific histological subtypes of breast cancer as its large size affords sufficient power for reliable comparisons. Complete follow-up for incident cancers and information on tumour histology is coded and routinely provided to the investigators by the NHS Central Registers. Such information has been demonstrated to be sufficiently valid to be used in epidemiological studies (


P=0.049) was found between fibre- and fat-tertiles. There was no significant association between breast cancer risk and intakes of any of the plant food subgroups. These findings support the hypothesis that a dietary pattern characterised by high fibre and low fat intakes is associated with a lower risk of postmenopausal breast cancer.The objective of this study was to investigate prospectively the associations between intakes of plant foods, fibre and relative fat and risk of breast cancer in a subsample of 11 726 postmenopausal women in the Malmö Diet and Cancer cohort. Data were obtained by an interview-based diet history method, a structured questionnaire, anthropometrical measurements and national and regional cancer registries. During 89 602 person-years of follow-up, 342 incident cases were documented. Cox regression analysis examined breast cancer risks adjusted for potential confounders. High fibre intakes were associated with a lower risk of postmenopausal breast cancer, incidence rate ratio=0.58, 95% CI: 0.40, 0.84, for the highest quintile of fibre intake compared to the lowest quintile. The combination high fibre–low fat had the lowest risk when examining the effect in each cell of cross-classified tertiles of fibre and fat intakes. An interaction ( Phyto-oestrogens especially have the potential to influence breast cancer risk. They are similar in structure to oestrogen and may compete with endogenous oestrogen for the oestrogen receptors  cohort have substantially lower intakes of fruit, vegetables and whole grain bread compared to low fat consumers and intakes of antioxidants and fibre are inversely associated with the relative fat intake  directed towards the community were used to recruit study participants. During the first 2 years, approximately 50% of the subjects were recruited by passive recruitment. However, during 1994–1996, less than 5% were passively recruited. Inadequate Swedish language skills and mental incapacity were the only exclusion criteria. When the baseline examinations closed in October 1996, 28 098 participants had completed all parts. More details of recruitment and the cohort are described elsewhere , migrating from Sweden (57 women) or end of follow-up (31 December 2001), whichever was the first. In total, 312 invasive and 30 in situ incident breast cancer cases were documented during 89 602 years of follow-up.The National Swedish Cancer Registry provided data until December 1999; additional information until end of follow-up (31 December 2001) was obtained from the Southern Swedish Regional Tumour Registry. Cases were women diagnosed during follow-up with invasive breast cancer or breast cancer The MDC method is an interview-based, modified diet history method. Briefly, it combines a 7-day menu book for recording of cooked meals, beverages, drugs, natural remedies and nutrient supplements and a diet questionnaire for assessment of meal pattern, consumption frequencies and portion sizes of regularly eaten foods. The mean daily intakes of food were calculated and converted to energy and nutrient intakes using PCKost2-93 from the National Food Administration in Uppsala, Sweden. The MDC method is described in detail elsewhere , fruit juice, vegetables, total fruit and vegetables, total fruit, vegetables and fruit juice, potatoes boiled, potatoes fried and deep fried, high-fibre bread, low-fibre bread, total bread, high-fibre cereals, low-fibre cereals, buns and cookies, nuts, snacks and rice and pasta.Total fat, n-6 polyunsaturated fatty acids (PUFAs), wine and fermented milk were also examined as potential confounders because these variables have been associated with breast cancer risk in previous studies in the MDC cohort (−1) was converted into a four-category variable based on the ranking of absolute intakes: wine-abstainers (zero consumption in menu book and no wine consumption during the last year indicated in the questionnaire); low consumers ; medium consumers .Energy-adjusted variables of total fat, n-6 PUFAs, fibre, total fruit and vegetables and fermented milk variables were defined as the residuals obtained when regressing the specific nutrient or food group on total energy intake (EI). Five exposure categories were created based on the quintile ranking of participants on residuals. In addition, for fat and fibre, a three-category variable was defined, based on the tertile ranking of participants on residuals. Information on wine consumption (cl dayMethod version (indicating data collection before or after 1 September 1994), diet interviewer and season of diet interview were used as covariates to control for undue variation in the dietary data collection over time.In September 1994, the dietary data processing procedure was slightly altered ; missing information (2.0%) was recoded to ‘no children’.Current use of hormone therapy was based on the question ‘which medicines do you use on a regular basis?’ in combination with the 7-day recording of drug use in the menu book, and used as a dichotomous (yes/no) variable , moderate (quintile 2–4) or high (quintile 5).Participants indicated the number of minutes per week, separately for the four seasons, for 17 different physical activities  between BMI and waist circumference, we decided to use only one obesity measure as a covariate. Waist was chosen because central obesity rather than general obesity has been suggested to predispose an individual to the development of breast cancer  was calculated. Basal metabolic rate was estimated using the equation recommended by WHO; based on age, sex, weight and height .T-test examined differences in the mean food and fibre intakes between cases and noncases.SPSS statistical computer package  was used for the statistical analyses. Crude median and interquartile intakes of food groups and fibre were calculated for cases and noncases. All dietary variables were log-transformed to reduce skewness. Before transformation, a very small amount (=0.01) was added to food intake variables in order to handle zero consumption. n=719, including 27 cases); women diagnosed with incident breast cancer within 1 year from baseline examination (31 cases) or in situ cancer cases (30 cases).The multivariate associations between food intake and breast cancer risk were examined with Cox regression models. Firstly, the influence of each of the 16 plant food-groups and dietary fibre on breast cancer risk was examined as continuous variables, while adjusting for diet interviewer, season of diet interview, method version, age at baseline, past food habit change and total energy (basic model). Food groups were further examined, in ‘full models’ also including established and potential risk factors . Secondly, the risk across quintiles of total fruits and vegetables and dietary fibre was estimated with both the ‘basic’ and the ‘full’ models. The ‘full models’ were extended to include adjustment for energy-adjusted quintiles of total fat, n-6 PUFAs, fermented milk or wine intake categories since these have been found to influence breast cancer risk in the MDC cohort. The models were also repeated while excluding women less than 51 years at baseline  and in the ‘full model’ (P=0.012).In the multivariate analysis, there were no significant associations between food group intakes (continuous) and breast cancer risk (data not shown). Fibre intake (continuous) was significantly associated with decreased risk, both in the ‘basic model’ .We found that high fibre intakes, quintile five compared with quintile one, was associated with a significant 40% reduction in breast cancer risk. The median fibre intake in quintile five was 25.9 g dayountries . The lowP=0.002) protective effect of fibre in a combined analysis of 12 case–control studies, but this effect was not independent of β-carotene and vitamin C. No association was found in the Nurse's Health Study . Although the combination high fibre–low fat had the lowest risk, there is no obvious pattern in breast cancer risk across the nine categories, which implies that other factors also might influence the observed risk estimates. There was no large difference in n-6/n-3 ratio across the nine categories. It varied from 4.99 in the medium fibre–high fat group to 5.25 in the medium fibre–low fat group. However, wine intakes varied substantially across the nine groups. There was a tendency of lower prevalence of high wine intakes with both increasing fibre intakes and increasing fat intakes. The prevalence of high wine intakes was 7.2% in the low fibre–low fat group and 0.0% in the high fibre–high fat group. Fermented milk intakes tended to decrease with higher fat intakes and increase with higher fibre intakes. The highest mean intake was found in the medium fibre–low fat group and the lowest mean intake in the medium fibre–high fat group. This is an interesting finding since it is possible that fermented milk influences the gut flora, which could lead to an increased activation of dietary lignans to active mammalian lignans . In factIn this study, we found no association between total fruit and vegetable intakes and breast cancer risk. The intakes of fruit and vegetables in the MDC cohort might be too low and too homogenous to examine any effect. As indicated by food intake data from the calibration study within the European Prospective Investigation into Nutrition and Cancer (EPIC), the MDC subcohort has among the lowest intakes of both fruit and vegetables in the EPIC study centres (Although epidemiological data support the apparent inverse association between consumption of fruits and vegetables and breast cancer risk, studies are not consistent overall . For insDuring the initial years of recruitment, over 50% of the participants were recruited by community-directed activities (Under-reporting of energy is a major concern when using self-reported dietary data . In thisTo conclude, a dietary pattern characterised by high fibre and low fat intakes was associated with a lower risk of postmenopausal breast cancer. The low intake levels of fruit and vegetables in the MDC cohort may contribute to the absence of an association between breast cancer risk and intakes of fruit and vegetables in this study.


The incidence of breast cancer in the unaffected breast of women with previous breast malignancy remains constant after the first diagnosis. We investigated whether there is a similar pattern in the breast cancer incidence in first-degree relatives of breast cancer patients. We studied the risk for breast cancer in mothers at ages older than their daughter's age at diagnosis.We analyzed a Swedish population-based cohort with complete family links and calculated incidence rates of breast cancer in mothers of 48,259 daughters diagnosed with breast cancer.The risk for breast cancer in mothers of breast cancer patients is elevated relative to the background population at all ages. Mothers have an overall incidence of 0.34%/year at ages older than a daughter's age at diagnosis. This rate is not affected to any large extent by the daughter's age at diagnosis. A constant incidence rate of 0.40%/year from age 35 years onward is seen in mothers of breast cancer patients diagnosed before 35 years of age. For mothers of daughters diagnosed at age 35 to 44 years the incidence pattern is less clear, with the rate being stable for approximately 20 years after the daughter's age at diagnosis and rising thereafter. Older age at a daughter's diagnosis (≥ 45 years) appears to confer an age-dependent increase in incidence in the mother.Incidence of familial breast cancer in first-degree relatives may increase to a high and constant level by a predetermined age that is specific to each family. This phenomenon appears inconsistent with accepted theories of malignant transformation. Studies of familial aggregation of breast cancer identify a family history of breast cancer as one of the strongest risk factors for the disease ,2. FamilYoung age at onset of disease within a family has long been regarded as a particularly strong risk factor for breast cancer -11. SeveFew studies have assessed the risk for disease on the absolute scale. Peto and coworkers  suggesteWe conducted a population-based study to characterize the rate of breast cancer in first-degree relatives at ages older than the index patient's age at diagnosis. In order to obtain a good age range for our study, we examined the risk for breast cancer in mothers of breast cancer patients.The Multi-Generation Register includes all Swedish residents born after 1931, who were alive in 1960, and all those born thereafter. It contains links between children and parents through their national registration numbers, which are assigned to all residents in Sweden. The register is updated yearly. During the period from 1961 to 2004 the completeness of the Multi-Generation Register became progressively better, and since 1991 it has been considered complete . ApproxiInformation on cancer was obtained from the nationwide Swedish Cancer Register, established in 1958. During the period of our study, the Cancer Register was estimated to be at least 98% complete [From a total population cohort comprising about 11 million individuals recorded in the Multi-Generation Register, we identified all women born in Sweden since 1932 with a first primary invasive breast cancer diagnosed during the period from 1961 to 2004. Subsequently, we identified all the mothers of these women, resulting in a total of 48,259 mother/daughter pairs in which the daughter had a breast cancer diagnosis. In cases in which more than one daughter from the same family had breast cancer, we randomly selected one of the mother/daughter pairs for inclusion in the study.The ethics committees of the Karolinska Institute  approved the study.Age-specific incidence rates in mothers of daughters with breast cancer were calculated using 5-year intervals. Mothers were studied up to age 85 years, and daughters could be studied up to age 73 years (born after 1931). For the analysis of complete follow up, person-years at risk began at the birth of the mother or 1 January 1961, whichever came later, independent of daughter's diagnosis. Mothers were followed only from the age at diagnosis of breast cancer in their daughter in order to estimate the incidence in mothers at ages older than their daughter's age at diagnosis. Person-years at risk ended on the date of diagnosis of a breast cancer, diagnosis of another malignant cancer, death, emigration or the closing date of the study (31 December 2004), whichever came first. Age-specific breast cancer risk in the general Swedish population (used for comparison) was estimated from the incidence of breast cancers in mothers from our database. The 'general population' was selected to be comparable to the mothers with affected daughters in terms of year of birth (1887 to 1961).P value of the test for linear trends in rates by maternal age categories. To investigate whether there is heterogeneity in linear trends of the rates in mothers that are approaching or have already passed the age at diagnosis of their daughters, we split the exposure time of mothers into two periods depending on whether the current age is younger or older than the age at diagnosis of the daughter. We then added to the previous model an interaction term between maternal age and this new variable (before/after). Likelihood ratio was used to obtain the P value of the heterogeneity in the trend estimates. In the same manner we tested heterogeneity in the trend estimates for mothers by age at diagnosis in daughters. All data preparation and analysis was done using the SAS statistical package, version 8.2 or higher  [Poisson regression was used to model the incidence of breast cancer in mothers for each 5-year age category. Rates were adjusted for calendar period in 5-year intervals. Likelihood ratio was used to obtain the NC, USA) .P = 0.0248). Mothers whose daughters were diagnosed before age 35 years had an increasing risk for disease up to the age of diagnosis in the daughter, and the risk remained roughly constant thereafter. Such a trend was not observed for older ages at diagnosis in the daughters. The mothers of daughters with breast cancer continued to be at increased risk (even at ages of 80 to 84 years) compared with women in the general population . Mothers of patients with a breast cancer diagnosis at age under 35 years exhibited a high and constant risk of about 0.40%/year at ages older than their daughters' age at breast cancer diagnosis. No increase in incidence rates by age was seen in this group of mothers, even after adjusting for calendar period using a Poisson model (P for trend = 0.32). Mothers with daughters diagnosed at age 35 to 44 years had a risk that appeared to be constant for approximately 20 years after the age at breast cancer diagnosis in their daughter and increased afterward. However, there was no significant difference in trend estimates for these two follow-up intervals (P = 0.44). Finally, incidence rates in mothers of daughters diagnosed at age 45 years or older increased steadily with age, with a significant trend (P < 0.05) after adjusting for calendar period.In our analyses of the age-specific incidence rates in mothers who have already reached the age at diagnosis in their daughters Table , we founvice versa are similar [Mothers of breast cancer patients are at increased risk for disease throughout the course of their life in comparison with the background population. At ages older than a daughter's age at diagnosis, her mother has an overall incidence that is not affected to any large extent by the age at diagnosis of her daughter. Mothers of breast cancer patients diagnosed at a young age are at high and constant risk from that age onward. We recognize that it is counterintuitive to consider breast cancer risk in mothers based on future breast cancer diagnosis in their daughters. However, we believe that the two individuals are interchangeable when discussing risk for breast cancer, primarily because the risk in daughters by mother and  similar ,12. We aThe strength of our study is in the prospective population-based design, large sample size, completeness of follow up and unbiased information on family history. Incidence rates for the general population are based on our database and are similar to the rates based on the Swedish Cancer Register .Although it would be more intuitive to study the risk for breast cancer in daughters or sisters of affected women, both of these analyses are subject to serious confounding between age and calendar date because of limitations in the data. Since the index women in our Multi-Generation Register were born after 1932, we could follow daughters only to a maximum age of 73 years, and any cancer in the oldest age group could only be diagnosed during the final years of follow up. As a result, it would be possible to begin follow up in sisters of older women only in recent years, again confounding calendar date and age. By studying the risk in mothers, not only can we follow women at all ages beyond the age at diagnosis of their relative (daughter), but also the diagnoses are evenly spread over the study period, with all ages represented in all years, allowing us to control for confounding by calendar date. Some mothers have been censored from our analyses, because the mothers could not be identified for 40% of women who died before 1991. However, we have shown elsewhere  that theUnilateral breast cancer rates are highly age dependent, being a rare event at young ages and more common as age advances, particularly after the menopause ,24. Our The most intriguing finding is the constant risk for breast cancer in mothers who have already attained the age of the index diagnosis in their family. This pattern was seen for mothers of breast cancer patients diagnosed before 35 years of age, with a constant rate for the whole follow-up period Table . For motAn age-independent model of familial breast cancer risk at ages older than the proband's age at diagnosis was previously proposed by Peto and coworkers . Those iBRCA1 or BRCA2, in whom the lifetime risks are quite substantial [The vast majority of familial disease is associated with a low penetrant polygenic aetiology that conveys lifetime risks somewhere below 30%, and only a small proportion are carriers of the highly penetrant mutations of stantial ,25-28. TThe observed patterns of familial rates of breast cancer suggest that age at onset might be inherited while the sporadic cancers dilute the message by superimposing a background risk in families where cancers are detected at older ages. This phenomenon seems inconsistent with accepted models of susceptibility and has implications for the carcinogenic process of breast cancer. Our findings suggest that screening of unaffected first-degree relatives should start at an age younger than the index patient's age at diagnosis. In the future we will probably be able to identify nonsporadic cancers, and it will become possible to direct screening activities toward these women.The authors declare that they have no competing interests.KC, MH, MR and PH contributed to the design. KC, MH, MR and PH were responsible for the presentation and implementation of this study. KC was responsible for data preparation and conducted the statistical analysis. KC and MH wrote the report. All authors contributed to the interpretation and discussion of the findings and approved the report.


We investigated whether elevation in serum cytomegalovirus (CMV) or Epstein–Barr virus (EBV) immunoglobulin G (IgG) antibody levels precedes the development of breast cancer.A nested case–control study was carried out within the Janus Serum Bank cohort. Two serum samples, one taken at least 4 years before diagnosis (sample 2) and an earlier sample (sample 1) from 399 women with invasive breast cancer and from 399 controls, matched for date of blood samples and age were tested for CMV and EBV IgG antibodies. Odds ratios (ORs) with 95% confidence intervals (CIs) for CMV and EBV seroconversion between the samples and unit changes in IgG optical density (OD) examined as a continuous variable were calculated using conditional logistic regression.Eleven cases and three controls seroconverted for CMV IgG between the first and second blood samples, with an adjusted OR for CMV IgG seroconversion of 4.0 (95% CI=1.1–14.4). The risk of breast cancer, adjusted for parity, increased per unit difference in CMV OD between samples . In an analysis restricted to parous cases and age-matched parous controls, the OR for CMV seroconversion for IgG between the two samples, adjusted for parity and age at first birth, was 9.7 (95% CI=1.2–77.3). The EBV seroconversion or change in EBV OD was not associated with risk of breast cancer.Our hypothesis that elevation in serum CMV IgG antibody levels precedes the development of breast cancer in some women is supported by the results of this study. Changes in EBV IgG antibody are not associated with risk of breast cancer. There are major geographical differences in the age-standardised incidence of breast cancer .Mouse mammary tumour-like viruses and Epstein–Barr virus (EBV) have been suggested to cause breast cancer, but the evidence that either is associated with breast cancer has been inconsistent .P<0.0001) was found between breast cancer incidence and the percentage of adults who are CMV seropositive (It has been hypothesised that breast cancer can be caused by late exposure to a common virus . Breast positive .Delayed exposure to EBV (or CMV), measured by illness from infectious mononucleosis, has been suggested to increase risk of breast cancer , P=0.005), but not for EBV . The adjusted odds ratios (ORs) per OD unit were 1.46 =1.06–2.03) for CMV IgG and 1.11 (95% CI=0.93–1.33) for EBV IgG. We hypothesised that the higher mean IgG levels found in women with breast cancer could be the result of more recent infection with CMV, and may indicate that late exposure to CMV (in adulthood rather than childhood) is a risk factor for breast cancer. Limitations of this work were that it was retrospective, with blood samples collected after the diagnosis of breast cancer, and only women aged <40 years were studied.In an Australian case–control study, cases and controls did not differ in seropositivity for CMV or EBV  and 10% are blood donors. The sera have been stored at −25°C .The stored blood samples from cases and controls were tested for CMV and EBV IgG antibodies. The CMV and EBV antibody levels in stored blood remain stable despite prolonged storage , and a blood sample at least 12 months earlier than the index sample. Eligible controls were women from the cohort who were alive and free of cancer  at the time that the case was diagnosed. They were frequency matched to the cases by 5-year age group and had a blood sample taken within ±2 months of the index sample of the case. From these eligible controls, women with at least one earlier sample were randomly selected. These controls were individually matched to cases by duration of sample storage (±2 months) of the earlier sample. Where it was not possible to find a control in which the earlier sample could be matched for duration of storage, an eligible control with the closest available early sample was randomly selected. The two samples for each case and individually matched control had to be separated in time by at least 12 months to allow adequate time for IgG levels to change between the samples, as CMV and EBV IgG titres rise initially after infection and then gradually decline, with residual antibody detectable for several years  and 800 from the controls (two samples from each of the 400 controls).On the basis of our earlier results  with similar proportions of nulliparous women, but fewer women with four or more live births among cases.Cases of breast cancer were diagnosed between February 1982 and December 2003, with 53% of cases diagnosed from 1998 onwards. One woman with DCIS only and her matched control were excluded, leaving 399 case–control pairs for analysis. The distribution of age and parity in cases and controls is shown in For the CMV IgG antibodies, the positive control provided with the kit had a mean OD value of 0.825±0.110 and the internal laboratory standard had a mean OD value of 0.235±0.052. For the EBV-VCA IgG antibodies, the positive control provided with the kit had a mean OD value of 2.610±0.230 and the internal laboratory standard had a mean OD value of 0.900±0.13.P=0.03). This effect was largely confined to parous women (10 cases and 2 controls). The mean CMV IgG for cases for sample 1 was 1.09 OD (s.d.=0.77) and the mean CMV IgG for controls was 1.18 OD (s.d.=0.71). The mean CMV IgG for cases for sample 2 was 1.19 OD (s.d.=0.79) and the mean CMV IgG for controls was 1.22 OD (s.d.=0.72). No significant differences in CMV IgG, with adjustment for parity, were found between cases and controls for sample 1 or 2 .The average time between serum samples was 8.5 years (range 1–17 years). Of the cases, 314 (78.7%) were seropositive for CMV at their second serum sample compared with 329 (82.5%) of the controls. Eleven cases and three of the controls seroconverted for CMV IgG between the first and second samples (The mean EBV IgG for cases for sample 1 was 2.08 OD (s.d.=0.80) and the mean EBV IgG for controls was 2.04 OD (s.d.=0.73). The mean EBV IgG for cases for sample 2 was 2.19 OD (s.d.=0.82) and the mean EBV IgG for controls was 2.17 OD (s.d.=0.80). With adjustment for parity, no significant differences in EBV IgG were found between cases and controls for sample 1 or 2 . For EBV, 386 (96.7%) of cases were seropositive when the second sample was taken compared with 384 (96.2%) of the controls. Two cases and three controls seroconverted for EBV IgG between the first and second samples .The CMV IgG levels did not appear to be affected by the duration of storage of the serum samples and no significant correlations between CMV IgG or EBV IgG level and duration of storage were found. There was no difference in the overall distribution of CMV IgG and EBV IgG levels in cases compared with controls.2=8.11, 4 df, P=0.09). Although a significantly increased risk of breast cancer was observed for parous women whose first birth occurred at age 28 years or older, no increased risk with increasing age at first birth with adjustment for parity was observed . Nulliparity did not confer reduced risk . The risk of breast cancer was significantly lower for women with five or more children compared with nulliparous women , with or without adjustment for age at first birth or CMV IgG seroconversion, and was similar with additional adjustment for BMI for the 277 case–control pairs for whom it was possible. In parous women, with adjustment for age at first birth, the risk of breast cancer decreased with increasing parity with an OR of 0.8 (95% CI=0.7–0.9) per additional child after the first and was not appreciably changed with additional adjustment for age at first birth.Seroconversion for CMV IgG between the first and second samples occurred for 11 cases and 3 controls, producing an unadjusted OR of 3.7 (95% CI=1.0–13.1). When adjusted for parity, the OR for CMV IgG seroconversion was 4.0 (95% CI=1.1–14.4). This was unaltered by adjustment for the length of time between tests, and was similar when restricted to case–control pairs for diagnoses before 1998 or 1998 onwards, OR=4.3 (95% CI=0.5–38.4) and OR=3.9 (95% CI=0.8–18.9), respectively. When only case–control pairs who were parous were included in the analysis (10 cases and 2 controls seroconverted), the unadjusted OR for CMV IgG seroconversion was 9.0 (95% CI=1.1–71.0), and this was increased slightly to 9.7 (95% CI=1.2–77.3) when adjusted for parity and age at first birth. The estimate was not altered significantly when the analysis was restricted to women over 50 years of age at diagnosis or reference date; OR=9.9 (95% CI=1.2–78.8).When the analysis was restricted to the subgroup for whom BMI was measured, the OR for the association between CMV seroconversion and breast cancer with adjustment for parity was 3.7 (95% CI=0.7–18.5) and 3.7 (95% CI=0.7–18.4) when adjusted for parity and BMI. No trend in the ORs over the four categories of BMI, with adjustment for parity, was observed . When restricted to parous women in this subgroup, the OR for CMV seroconversion was 7.6 (95% CI=0.9–64) and was unchanged with additional adjustment for age at first birth.Changes in CMV serology from positive to negative, or staying positive or staying negative, between samples were not associated with risk of breast cancer. Only 2 cases and 3 controls seroconverted for EBV IgG between the two samples.The crude OR per unit of difference in CMV OD between samples was 1.6 (95% CI=1.0–2.3), and the crude OR per unit difference in EBV OD between samples was 0.9 (95% CI=0.7–1.3). Our hypothesis that elevation in serum CMV or EBV IgG antibody levels precedes the development of breast cancer in some women is supported by the results of this study. Seroconversion of CMV IgG and increasing IgG levels between the samples were both associated with an increased risk of breast cancer among parous women.The Janus serum bank cohort was ideal for this study because it meant that blood, which was collected years before the diagnosis of breast cancer, could be tested, so CMV and EBV IgG levels for cases are unlikely to have been affected by breast cancer and were not affected by treatment. Exposure to CMV and EBV was determined objectively using standard enzyme immunoassays for CMV IgG and EBV viral capsid antigen IgG, and laboratory staff were blinded to the case or control status of the samples, so these results are not vulnerable to recall bias or observer bias. The similar risk for case–control pairs before and after 1998, the year the Norwegian Breast Cancer Screening Programme began, suggests that the results were not affected by detection bias associated with screening.Adjustment for the length of time between samples did not alter the results obtained and suggests that recent seroconversion rather than the duration of seropositivity may be the more important determinant of risk of breast cancer. Whether CMV remains in breast tissue after infection in human beings is not known. Increased CMV IgG levels may occur from recrudescence of latent infection or re-exposure to CMV without clinical infection. Increasing risk of breast cancer with increasing CMV OD might also occur if a significant risk factor for breast cancer was closely associated with recurrent CMV exposure.Day care workers and mothers are at increased risk of acquiring CMV from children they care for (The risk of adult CMV infection might be expected to increase with increasing parity, but increasing parity is associated with a reducing risk of breast cancer. However, if the prevalence of pre-invasive disease increased with age and exposure to CMV only promoted the progression of pre-invasive disease, then infection from young children after an early age of first full-term birth or increasing parity would provide some immunity against infection at an older age, after the onset of pre-invasive disease, and thus reduce the risk of breast cancer. Then both early age of first birth and increasing parity would, as has been observed, reduce the lifetime risk of breast cancer. Moreover, the increased risk of breast cancer observed within 10 years of first full-term birth (Studies involving the long-term follow-up of women who have been, and not been, infected with CMV are required to verify the findings of this study. A greater time period between blood samples would be expected to increase the proportion of women for whom CMV IgG seroconversion occurs and provide more definitive examination of the association with breast cancer and whether risk increases with increasing age of adult CMV infection. Assessment of whether the increased risk of breast cancer in the 10-year period after first full-term birth varies by occurrence of CMV infection during the period is also needed.


Those having a child with a childhood malignancy had a significantly worse survival than other women, relative risk (RR)=1.25, 95% CI 1.02–1.55, P<0.04, adjusted for age at diagnosis, year of diagnosis, parity and time since last pregnancy. Childhood sarcomas or acute myeloid leukaemia seemed to be most associated with a worse survival in the mother . The lower survival of the mother was present for breast cancer diagnosed both before and after 50 years of age. The Li–Fraumeni syndrome and possibly other genetic disorders may lower breast cancer survival.As it is unclear if hereditary factors affect breast cancer survival, this was compared using fertility and cancer registry data, among all women so diagnosed during 1961–1999 in Sweden, having a child with childhood cancer (⩽20 years of age;  Only a minority of cases of breast cancer have a hereditary background. Hereditary factors could affect tumour biology and disease prognosis and it has been suggested that breast cancer patients having p53 germline mutations have an especially bad prognosis , 2001. In=75 035) together with deaths, emigrations and tumours in their children. Only primary tumours were considered and no information about recurrences or development of secondary tumours is available. The women were followed from breast cancer diagnosis to the first event of emigration, death or 31 December 2001.Women with at least one child and a breast cancer diagnosis during 1961–1999 were identified by means of the unique personal identification numbers and the National Swedish Cancer register, fertility register and Swedish national censuses . Risk was highest if the child had acute myeloid leukaemia  or sarcoma . Adjusting for number of children and time since last childbirth only marginally changed the risk estimates. Censoring the follow-up of all women at 65 years of age did not influence the results.As shown in In this population-based national registry study, a worse breast cancer survival was found for women also having an offspring with a paediatric malignancy compared to other parous women not having a child with malignancy. The most plausible explanation is that a genetic factor, such as p53 germline mutations, in a small subgroup of the first-mentioned women, also affects tumour biology causing the worse survival, as we postulated before . Breast In this registry study, we lacked information about tumour stage or other biological factors related to prognosis and about therapy. We were, however, able to adjust for age at diagnosis, birth year, number of children and time since last childbirth, as breast cancer diagnosed in pregnancy or in its close proximity has an especially poor prognosis  but would have no major effect in younger women. Thus, the smaller effect seen for older patients could be due to differences in intercurrent deaths.There is no reason to suspect inclusion bias that an offspring having childhood cancer is related to the likelihood of a woman being identified by the registries. Most childhood cancers occurred before the breast cancer diagnoses in mothers, but available evidence gives little support to the notion that a stress reaction or grief can affect the prognosis of breast cancer. However, an alternative hypothesis is that the shared environmental effects might also contribute to an increased mortality risk.Further studies with tumour stage, hormone receptor status, nuclear grade, erbb2 status, therapy and p53 germline mutation status may elucidate this link between childhood cancer and breast cancer. Our results suggest that data on whether a woman has had an offspring with childhood cancer have prognostic value.


For participants in both arms of the trial, there were improvements in subjective and objective understanding up to 4 weeks which were generally sustained up to 6 months. However, improvements in subjective understanding for the women at low risk of breast cancer (i.e. not at significantly increased risk) in the standard service arm did not reach statistical significance. Cancer worry was significantly reduced at 6 months for participants in both arms of the trial. The two models of cancer genetics services tested were generally comparable in terms of the participants' psychological outcomes. Therefore, decisions regarding the implementation of the novel community-based service should be based on the resources required and client satisfaction with the service.This study compared the psychological impact of two models of breast cancer genetics services in South East Scotland. One hundred and seventy general practices were randomised to refer patients to the existing standard regional service or the novel community-based service. Participants completed postal questionnaires at baseline ( Media attention to scientific developments in cancer genetics has resulted in a greatly increased demand for cancer genetics services. These services aim to identify individuals who have inherited a significantly increased risk of cancer in order to counsel them about their risks and to offer appropriate risk management to reduce morbidity and mortality. There is a challenge to provide this information in ways that the lay public can utilise to inform their health-care choices without causing undue psychological distress. Individuals who are not at significantly increased risk also need appropriate reassurance without precluding an appropriate vigilance to symptoms of sporadic cancer. There is also a challenge to respond to these new developments within existing health-care budgets. Internationally, there is a lack of consensus about how best to deliver cancer genetic services .An alternative model of cancer genetics services has been proposed , South West Fife (n=54) and Borders (n=24) Health Boards in South East Scotland. One hundred and seventy practices (84%) agreed to take part, 23 (11%) declined and 10 (5%) did not reply. This meant that 725 of the 828 (88%) GPs in practice across these three Health Boards agreed to refer patients into the trial. Practices were randomly assigned to either arm of the trial using a minimisation technique  and genetics nurse specialist (JC) assigned categorical risk assessments informed by published criteria , run by a genetics nurse specialist (RC/RT). At the clinic, the genetics nurse specialist ascertained the woman's family history of cancer and compiled a family tree. This information was compared to published criteria  to deterWomen were asked to record their date of birth, marital status and educational level on the baseline questionnaire. Information about the category of breast cancer risk to which each woman had been assigned was derived from the clinical records.not at all) to 4 (very much) how well they understood each of four issues relevant to breast cancer genetic risk. The issues were:How increased risk of breast cancer is passed on in families.The significance of their own family history of cancer.Whether there was anything they could do to reduce their risk of developing breast cancer.What services were available to protect the health of people at increased risk of breast cancer.Women were asked to rate on a 4-point scale from 1 (Responses were summed to give a composite score for subjective understanding ranging from 4 to 16.correct) or 0 (incorrect/don't know), and the number of correct responses combined to give total scores for genetics understanding (range 0–10) and mammography understanding (range 0–12).Participants were asked to consider a number of factual statements and to respond ‘true’, ‘false’ or ‘don't know’. There were 10 statements about breast cancer genetics (e.g. ‘Only a parent who has had breast cancer can pass on increased risk to their children’) and 12 statements about issues surrounding mammography (e.g. ‘Mammograms are used to detect early stages of breast cancer’). Statements were scored 1  ((GHQ-30) . This weCancer Worry Scale . They were asked to rate whether the frequency of any of these behaviours had changed since consulting genetics services (at 4 weeks) or in the last 6 months (at 6 months) on a scale from 1 (much less than before) to 5 (much more than before).Several t-tests (two-tailed), Mann–Whitney, χ2 (two-tailed) or Fisher's exact tests (two-tailed). A 2  × 2 (objective risk) repeated measures analysis of variance (ANOVA) was used to determine between-group differences and within-group changes over time  in psychological outcomes and possible interactions between trial arm, objective risk and time. Significant effects were followed up with post-hoc tests . χ2 (two-tailed), Fisher's exact (two-tailed), Cochran's and McNemar tests were used to examine the impact of time, trial arm and objective risk on perceived risk and the proportion of participants suffering from ‘case-level’ distress. A significance level of 0.05 was used throughout. The data were analysed using SPSS for Windows version 10.00 (1999).Descriptive statistics were generated to describe the study participants. Differences between two independent groups were analysed with independent samples Figure 1Over the study period, 574 women, referred for breast cancer genetic risk counselling, were invited to participate in the trial. Consent forms were returned by 451 women (response rate 79%), of whom 428 (75% of those invited) agreed to participate in the study. Three hundred and seventy-three of these women (87% of those who consented) returned a completed baseline questionnaire, 185 of whom were then assigned to the standard service arm and 188 to the novel service arm of the trial according to their GP practice.Of the 373 women who completed a baseline questionnaire, 276 also completed a 4 week follow-up questionnaire , 147 from the standard service arm and 129 from the novel service arm of the trial.n=97) were compared with those of the ‘4 week group’ (n=276) to check for participation bias. A significantly greater number of the ‘baseline only group’ had been assigned to the novel service arm than the standard service arm of the trial . A higher proportion of women in the ‘baseline only group’ were categorised as being at low risk . Similarly, a greater proportion of women in this group were suffering from ‘case-level’ distress at baseline . The ‘baseline only group’ had significantly higher scores at baseline on the Cancer Worry Scale . There were no significant differences between the two groups on any of the other sociodemographic or psychological variables at baseline.The characteristics of those for whom only baseline data were available  or nonreturn of the questionnaire (n=10). The baseline characteristics of those for whom the 6 month questionnaires were analysed  were compared with those participants who completed the 4 week questionnaire but not the 6 month questionnaire  to check for participation bias. There were no significant differences between the number of women who dropped out from either arm of the trial. A greater proportion of women in the ‘Not 6 month’ group were classified as being at low risk of breast cancer . This group scored significantly lower on objective understanding of mammography  and had a tendency to have higher scores on the Cancer Worry Scale . There were no significant differences between the two groups on any of the other sociodemographic or psychological variables .Two hundred and sixty-three women completed 6 month follow-up questionnaires , 140 women from the standard service arm and 123 from the novel service arm of the trial. This includes 19 women who were excluded at the 4 week assessment due to administrative reasons , 4 weeks  and 6 months . Women with 4 week follow-up data in the novel service arm were somewhat older than those in the standard service arm . There were no significant differences between the trial arms on any of the other sociodemographic variables at the three assessment points.The sociodemographic characteristics of participants and the breast cancer risk category to which they were assigned are shown in Table 2Table 3t=−14.97, df=231, P<0.000). Scores on subjective understanding were shown to be dependent on objective breast cancer risk. Post-hoc analysis revealed that women at moderate/high risk had significantly greater scores on subjective understanding than those at low risk at 4 weeks  and 6 months . The main effects of time and objective risk were modified by a significant interaction between these two factors. Subjective understanding had significantly improved for both women at moderate/high risk  and women at low risk  between baseline and 4 weeks. However, the improvement was significantly greater in women at moderate/high risk than those at low risk . In addition, there was a significant interaction between trial arm, time and objective risk. Post-hoc analysis indicated that differences in subjective understanding between the risk groups within the different trial arms were significant only between baseline and 4 weeks =5.27, P=0.023). Between these two time points, the moderate/high-risk women in the standard service arm  and low risk  and moderate/high-risk women  in the novel service arm had made significant improvements in subjective understanding. There were no significant differences in the extent to which subjective understanding had improved between these groups. Although for women at low risk in the standard service arm there was an improvement in subjective understanding between baseline and 4 weeks, this did not reach statistical significance.Overall, there was a significant improvement in subjective understanding during the course of the study. Further analysis showed that subjective understanding only improved to a significant degree between baseline and 4 weeks .There was a significant improvement in objective understanding of genetics and mammography across all participants during the study period. Post-hoc analysis revealed that the greatest reduction in scores occurred between baseline and 4 weeks  with a smaller, but nevertheless significant reduction between 4 weeks and 6 months .For all participants, there was a significant reduction in scores on the Cancer Worry Scale during the course of the study. Q=11.44, df=2, P=0.003), further investigations showed that the reduction was only significant between baseline and 4 weeks . There were no significant differences in the proportion of women suffering from ‘case-level’ distress between trial arms or risk groups at the three assessment points.Although there was a significant decrease in the overall proportion of participants suffering from ‘case-level’ distress over the study period . Further analysis showed that these changes were only significant between baseline and 4 weeks where significantly less women perceived their risk as low at 4 weeks (P=0.011). There were no significant differences in perceived risk by trial arm at the three assessment points. However, a significantly greater proportion of women at low objective risk of breast cancer than those at moderate/high objective risk perceived their risk to be low at 4 weeks  and 6 months .There were significant changes in perceived risk of breast cancer across all subjects over the study period (Cochran's vs 23%) and proportionately more women in the novel service arm reported examining their breasts more frequently than once per month . There were no significant differences between the two trial arms in the extent to which participants reported performing health behaviours prior to genetic counselling or reported a change in these behaviours after counselling. At 6 months, there were no significant differences between the two groups in the proportion of women who reported changing any of their health behaviours in the last 6 months.At 4 weeks, proportionately more women in the standard service arm reported examining their breasts every month as recommended  with 65% of those enrolled in the trial completing all three questionnaires. The amount of data lost due to administrative reasons was comparable across the trial arms.Women who dropped out of the study tended to be in the novel service arm of the trial or at low risk of breast cancer. The latter finding is not unexpected since these women may have been less motivated to continuing participating in a study of cancer genetics services which they were ineligible to receive. However, the women who dropped out of the study had greater levels of psychological distress. As these women may have dropped out in an effort to reduce their high levels of distress, they could perhaps benefit from further psychological intervention. Similar findings have been demonstrated by a previous trial of cancer genetics services  and this was reflected in the objective assessment of understanding. On average, correct responses were given to about one-third of the breast cancer genetics items and to about two-thirds of the mammography items. About one-third of participants were suffering from ‘case-level’ distress. This is comparable to the findings in other samples of women prior to genetic risk counselling using the same measure and threshold  and to pThe findings show that after consulting cancer genetics services, many of the short-term improvements in psychological outcomes experienced across participants were maintained up to 6 months. All participants reported greater subjective understanding of issues related to breast cancer risk and these improvements were most marked up to 4 weeks and were generally sustained up to 6 months. However, for the women at low risk of breast cancer in the standard service arm of the trial, unlike all other participant groups, these improvements did not reach statistical significance. This may be due to the fact that the majority of these women received a letter informing them about their low risk and were not offered a face-to-face consultation. Improvements in subjective understanding across participants were reflected by improvements in objective understanding which were again most evident between baseline and 4 weeks and were commonly maintained at a similar level up to 6 months. Although participants at low risk of breast cancer in the standard service arm did not feel that their subjective understanding had improved as much as the other participants, there were no differences between trial arms or risk groups in significant improvements on objectively assessed understanding.It was reassuring to find that despite improvements in objective knowledge, the proportion of women suffering from ‘case-level’ distress decreased up to 4 weeks and cancer worry continued to decrease up to 6 months. Unlike previous research that found significant reductions in cancer worry only for those women at low or moderate risk of breast cancer (i.e. ad-hoc items were analysed for the purposes of this report and the accuracy of participants' risk perceptions was not assessed in this study, it is difficult to make any firm conclusions from these results. However, there is no evidence to suggest that learning that your risk of developing breast cancer is greater than you believed prior to genetic risk counselling causes psychological distress (Women's perceptions of their risk of breast cancer were altered during the course of the study with significantly fewer women overall perceiving their risk as low at 4 weeks, than at baseline. More women who were informed about a low risk of breast cancer perceived their risk of breast cancer as low following their risk assessment, as may be expected. Similar findings have been reported elsewhere (This study has shown that the novel community-based model of breast cancer genetics services is generally comparable to the existing standard regional service in terms of the psychological outcomes experienced by recipients. Therefore, decisions regarding the implementation of the novel model of services should be based on additional factors such as the resources required and client satisfaction with the service. These factors have been investigated and will be published separately.