CCND1) plays a key role in cell cycle control, particularly in the transition from G1 to S phase, which is regulated by cyclin-dependent kinases. A common adenine to guanine polymorphism (A870G) in the CCND1 gene has been associated with a longer-life protein and an increased risk of colorectal cancer and adenoma in some studies. Among subjects with hereditary nonpolyposis colorectal cancer, the A870G polymorphism has also been associated with a younger age of onset of colorectal cancer. We analysed 181 colorectal cancer cases and 475 matched controls and 524 adenoma cases and 517 matched controls within women in the Nurses’ Health Study (NHS) cohort, 171 colorectal cancer cases and 347 matched controls and 372 adenoma cases and 712 matched controls nested within men in the Health Professionals’ Follow-Up Study (HPFS) cohort, and 258 colorectal cancer cases and 415 matched controls within men in the Physicians’ Health Study (PHS) cohort to assess the risk associated with the CCND1 A870G genotype. Moreover, we assessed whether CCND1 genotype modified the effect of a sporadic (nonsyndromic) family history of colorectal cancer as well as the effect of other dietary and lifestyle risk factors for colorectal cancer and adenoma. In all cohorts combined, the CCND1 polymorphism did not show statistically significant associations to risk of colorectal cancer (odds ratio (OR) for A allele carriers, 1.04; 95% confidence interval (95% CI), 0.82–1.32) or adenoma . The CCND1 A870G genotype was associated with a modest, although nonsignificantly elevated risk of colorectal cancer  in women. In contrast, the polymorphism was not associated with increased risk of adenoma in either men or women. Among participants with the A870G genotype, a family history of colorectal cancer conferred a substantially greater risk of colorectal cancer in the women (P for interaction=0.06) and adenoma in the men (P for interaction=0.02). Current postmenopausal hormone (PMH) use was associated with a significant reduction in the risk of colorectal cancer and adenoma among women with the A870G genotype, whereas there was no effect of PMH use among those with the GG genotype. The CCND1 polymorphism appeared to confer a modest elevation in the risk of colorectal cancer among women. Moreover, the A870G genotype may enhance the protective effect of postmenopausal oestrogen use on the development of colorectal neoplasia.Cyclin D1 ( CCND1) is a protein that plays a key role in cell cycle control, particularly in the transition from G1 to S phase, which is regulated by cyclin-dependent kinases  (CCND1 genotype.Previous studies have had inconsistent results for an association of the enotype)  was assoCCND1 A870G polymorphism on the risk of colorectal cancer and adenoma in three large prospective cohort studies, the Nurses’ Health Study (NHS), the Health Professionals’ Follow-Up Study (HPFS), and the Physicians’ Health Study (PHS). Using banked blood specimens and regularly updated data on dietary and lifestyle risk factors, we assessed whether the influence of these environmental factors was modified by the CCND1 polymorphism.We therefore examined the influence of the The NHS is an ongoing prospective study of 121 700 US female registered nurses. Details of the design and follow-up of this cohort have been described previously . As in previous studies  of the randomised physicians. The men were subsequently followed for incident cancer through annual mailed questionnaires. By the year 2000, 272 cases of colorectal cancer were identified and confirmed using medical records. Men who were free from diagnosed cancer at the time of cancer ascertainment (n=456) were selected as controls, and were matched on age  and on smoking history at baseline .The PHS is a randomised, double-blind trial of aspirin and μl buffy coat diluted with 150 μl of phosphate-buffered saline and using the QIAmp  96-spin blood protocol according to the manufacturer's instructions. Genomic DNA concentrations were calculated in 96-well format using PicoGreen technology .In both the NHS and HPFS, venous blood samples were separated into plasma, buffy coat and red blood cells and stored in liquid nitrogen. Genomic DNA was extracted from whole blood samples (PHS) or from 50 CCND1 was carried out using the TaqMan allelic discrimination system . TaqMan primers and probes are available upon request from the authors. Following polymerase chain reaction amplification, end point fluorescence was read with the Applied Biosystems 7900HT instrument and genotypes were assigned using Allelic Discrimination Software (Applied Biosystems SDS Software v.1.7a). Quality control  samples were included and each analysis included no DNA template controls. Laboratory personnel were blinded to QC and case–control status. In all, 1697 women from the NHS  were successfully genotyped, as well as 2275 men from the HPFS  and from the PHS .Genotyping of CCND1 genotypes. Cyclin D1 A870G was categorised into the three genotypes  and as noncarriers  vs carriers of the variant A allele . We conducted analyses for all three cohorts separately and then combined analyses, pooling data from the NHS, HPFS, and PHS, with adjustment for age, gender, and body mass index (BMI). For analysis of the main effect of genotype, logistic regression was used to compute odds ratio (OR) and 95% confidence intervals (CIs), control for a variety of potentially confounding variables, and to test for gene–environment interactions. Although the data sets were initially based on matched cases and controls, the stratified analyses required unconditional analysis. Moreover, conditional logistic regression yielded similar results as obtained when running unconditional logistic regression models, thus we report OR) derived from unconditional regression models throughout the manuscript. Specifically, analyses involving NHS, HPFS, and PHS cancer cases and controls were controlled for age, family history of colorectal cancer (NHS and HPFS only), smoking history, aspirin use, BMI, postmenopausal status and postmenopausal hormone (PMH) use (NHS only), physical activity, and intake of red meat, folic acid and alcohol, and history of previous endoscopy (HPFS only). For colorectal adenoma, separate analyses for men and women using unconditional logistic regression were controlled for age, history of previous endoscopy, year of endoscopy, family history of colorectal cancer, smoking history, aspirin use, BMI, PMH use among postmenopausal women (NHS only), physical activity, and intake of red meat, charred meat, folate, and alcohol. In subanalyses, men and women in the adenoma groups were combined for gene–environment interactions and were controlled for age, sex, and family history of colorectal cancer. Analyses involving PHS cancer cases and controls were controlled for age, smoking status, aspirin use, BMI, physical activity, and intake of red meat, vitamins, and alcohol.Adenoma risk was considered in relation to the CCND1 genotype. In the NHS colorectal cancer group, CCND1 genotypes were analysed in combination with micrograms (μg) per day of folate including supplements , milligrams (mg) per day of vitamin B6 pyridoxine  and grams (g) per day of methionine , based on median distributions in the control population, regular aspirin intake , family history of colorectal cancer , pack-years of smoking before age 30 , g per day of alcohol , age at diagnosis , and PMH use among postmenopausal women . In subanalyses, we stratified on the stage of disease (stage I (muscle only) vs stages II and III combined ) and site (colon vs rectum). In the NHS adenoma group, CCND1 genotypes were analysed in combination with μg per day of folate including supplements , mg per day of vitamin B6 (pyridoxine), and g per day of methionine , based on median distributions in the control population, family history of colorectal cancer  both alone and in conjunction with adenoma size , pack-years of smoking , g per day of alcohol , age at diagnosis , regular aspirin intake , PMH use among postmenopausal women , site , and size  of adenoma. In the HPFS adenoma group, CCND1 genotypes were analysed in combination with μg per day of folate including supplements , and g per day of methionine , based on median distributions in the control population, regular aspirin intake , age at diagnosis , pack-years of smoking , g per day of alcohol , and family history of colorectal cancer . In subanalyses, we stratified on age at diagnosis  and site , and size  of adenoma. In the HPFS cancer group, CCND1 genotypes were analysed in combination with μg per day of folate including supplements , and g per day of methionine , based on median distributions in the control population, regular aspirin intake , age at diagnosis , pack-years of smoking , g per day of alcohol , and family history of colorectal cancer . In subanalyses, we stratified on the stage of disease (advanced vs not) and site (colon vs rectum). In the PHS colorectal cancer group, CCND1 genotypes were analysed in combination with smoking , vitamin intake , aspirin assignment , age at diagnosis , site (rectum vs colon), alcohol intake in drinks per day , and age . Exposure information for the NHS was updated from 1990, 1992, 1994, 1996 to 1998. Exposure information for the HPFS and PHS was collected at baseline in 1986 and 1982, respectively.The effects of various dietary and lifestyle exposures on colon cancer and adenoma risk were tested in conjunction with P-values are two-sided. All statistical analyses were carried out using the SAS 8.2 statistical package .Statistical tests for interaction was based on the Wald test for the crossproduct term in a model containing the main effects of genotype and exposure variable as continuous variables. All A recent case–control study , 1994b. CCND1 A allele frequency ranged from 53 to 56% in the NHS, HPFS, and PHS control populations and the genotype distributions were in Hardy–Weinberg equilibrium. The A allele frequency was slightly more frequent in our populations than in similar, white populations of previous reports (42–43%) , 1.04; 95% CI, 0.82–1.32) or adenoma . Moreover, the risk associated with the homozygous AA genotype did not appear to differ materially from the heterozygous GA genotype . In contrast, the variant genotype was not associated with risk of cancer or adenoma in men or risk of adenoma in women. In addition, CCND1 genotype did not confer a significant elevation in the risk for either large (⩾1 cm) vs small (<1 cm) adenoma in either cohort (data not shown).In all cohorts combined, the genotype . The CCNCCND1 genotype varied for proximal as compared to distal colorectal cancers. The CCND1 A870G polymorphism appeared to confer an elevated risk for cancers in the distal colon or rectum (rectum to splenic flexure) , but not for proximal cancers (splenic flexure to caecum)  in the women. In contrast, among men in the HPFS, we found similar risks for both proximal  and distal cancers . Data on cancer location within the bowel were not available from the PHS cohort. Previous studies reported a younger age of onset for colorectal cancer for subjects with HNPCC who possessed the AA or GA CCND1 genotype or the presence of the variant truncated CCND1 transcript   on cancer and adenoma risk (P for interaction=0.06). However, in this cohort of women, we found no material interaction between genotype, family history and adenoma risk.We assessed whether the influence of enotype) . We theroma risk . Among fCCND1 genotype on the relation between family history and adenoma risk in the HPFS was statistically significant (P for interaction=0.02). Whereas family history conferred no significant influence on adenoma risk among men with the GG genotype, family history was associated with a significantly increased adenoma risk  among men with the GA or AA genotypes. Nonetheless, among men in the HPFS, we found no significant interaction between genotype, family history, and colorectal cancer risk (data on family history were not available from the PHS cohort). In addition, we examined whether age at onset of disease influenced the overall association between the genotype and colorectal cancer or adenomas. In these analyses, we observed a higher risk of colorectal cancer among women who carried one or two copies of the A allele and who had a diagnosis of colorectal cancer at a later age . Similar risk increases were not observed for colorectal adenomas among women, or among men.Similarly, among those with a family history, there was an increased risk from the GA or AA genotype, compared to those with the wild type. Among men in the HPFS, the influence of CCND1 genotype. In analyses of both cancer and adenoma, the influence of BMI, smoking, aspirin use, and intakes of folate, red meat, and alcohol did not differ significantly according to CCND1 genotype . However, among postmenopausal women, the inverse relation between postmenopausal oestrogen use and colorectal cancer and adenoma risk appeared to be limited to the A870G polymorphism  , thus potentially limiting the generalisability of our adenoma findings. Moreover, since we limited our analysis to adenoma within the reach of the flexible sigmoidoscope, we were unable to assess the association of genotype with the risk of proximal adenoma. However, for the analysis of colorectal cancer in our cohort of women, the risk associated with A870G genotype was, in fact, greater for cancers of the distal colon and rectum than for more proximal lesions. Finally, even though our study is of fairly large size, it may still have been underpowered for certain gene–environment interactions.CCND1 A870G genotype. Moreover, the A870G genotype appears to enhance the protective effect of postmenopausal oestrogen use on the development of colorectal neoplasia. Additional studies are warranted to confirm this potential interaction between CCND1 genotype and postmenopausal oestrogen use.In summary, our results suggest that the risk of both colorectal cancer and adenoma associated with a family history of colorectal cancer may be augmented by the


We evaluated the TGFBR1 9A/6A and Int7G24A variant frequencies in two breast cancer cohorts; a population-based cohort of breast cancer with defined family history (n=459) and in breast cancer patients from a familial cancer clinic (n=340) and in 856 controls from the Stockholm region. The familial patients from both cohorts were further divided into high- and low-risk familial breast cancer based on pedigree analysis. There was no overall association with either variant and breast cancer risk. The TGFBR1*6A allelic frequency was, however, higher in low-risk familial breast cancer (0.138), compared to controls . No significant difference was found in the high-risk familial (0.102) or sporadic cases . TGFBR1*6A carrier status was further associated with a high-grade sporadic breast cancer . These results indicate that the TGFBR1*6A variant may be associated with an increased risk of low-risk familial breast cancer and might be a marker for poorly differentiated breast cancer. The Int7G24A variant was not associated with breast cancer risk or clinical presentation of the disease including prognosis in our material.Two common variants in transforming growth factor- BRCA1, BRCA2, p53, ATM and PTEN, account for less than 25% of the familial risk for breast cancer, and in the Stockholm region the frequency of mutations in BRCA1 and BRCA2 is less than 10% in families with three or more cases with breast cancer  signalling is involved in the development of various types of cancer. As one of the key effectors of TGF-β signalling, TGF-β receptor 1 (TGFBR1) mediates the growth-inhibitory signals from TGF-β through a complex with TGFBR2. Growing epidemiological evidence indicates that common variants of the TGF-β pathway receptors that alter TGF-β signalling can modify cancer risk   from October 1998 to May 2000 were asked participate in a study on genetic risk factors from breast cancer . In the cases from the Department of Oncology, the mean age was 60 years (27–88 years), and there was no statistically significant difference between familial and sporadic cases .As controls, we used DNA from 856 blood donors of mixed gender collected at Karolinska University Hospital, Stockholm, Sweden.The ethical committee at the Karolinska Institute approved of the study.μl containing 50 ng DNA and 1.25 U Platinum® Taq DNA polymerase (Invitrogen). The fluorescently labelled PCR products were then separated on an ABI 377 DNA Sequencer . The genotypes were analysed using the GENESCAN™ and GENOTYPER™ software (Applied Biosystems). A product size of 247 bp represented the wild-type TGFBR1 (9A) allele, whereas a product size of 255 bp represented the TGFBR1*6A allele.PCR amplification of exon 1 was performed with the following primers: Fwd-5′-FamGAGGCGAGGTTTGCTGGGGTGAGG-3′ and Rev-5′-CATGTTTGAGAAAGAGCAGGAGCG-3′. The reactions were performed using supplied protocol for GC-rich fragments  in a total volume of 25 μl containing 50 ng DNA and 0.625 U AmpliTaq Gold® DNA polymerase (Applied Biosystems). A total volume of 10 μl PCR product was digested using 3 U of Bsr1 restriction enzyme in 1 × NEBuffer3 at 65°C for 8 h. The digested DNA products were then separated in a 2.5% agarose gel and visualised by ethidium bromide staining. Bsr1 cuts the wild-type Int7G24A sequence while the Bsr1 restriction site is eliminated by the variant allele. For all identified variant carriers and for approximately 25% of the wild-type samples, restriction enzyme digestion was replicated after a second independent PCR amplification to confirm the allele calling. The concordance rate was 100%.Genotyping primers for the Int7G24A variant were Fwd-5′TGTCTGAAAGGAGGTTCATCC-3′ and Rev-5′-GAACAACTTCTGCTCATGACG-3′. The PCR was carried out in a total volume of 25 χ2-test. Genotype distribution in breast cancer cases and controls was tested for the Hardy–Weinberg equilibrium, and the two variants were shown to be in the Hardy–Weinberg equilibrium in both cases and controls.The allelic frequency was analysed by the Genotype data were compared between the groups using univariate logistic regression analysis for binomial and ordinal responses. For the ordinal response variable grade, we used a model where each non-reference category was contrasted with the reference category (grade 1). Results are presented as odds ratios (ORs) with 95% test-based confidence intervals (CIs). No correction for multiple statistical testing was performed.*6A genotypes were scored for 95.6% (764 out of 799) and 100% (856 out of 856) of the cases and controls, respectively, corresponding to genotype results for 309 sporadic, 254 low-risk and 201 high-risk familial cases and 856 controls. The Int7G24A variant was successfully genotyped in 96.0% (767 out of 799) of the cases and 99.6% (853 out of 856) of the controls, with genotype results for 311 sporadic, 250 low-risk and 206 high-risk cases and 853 controls.TGFBR1P=0.17) nor in the population-based cohort (P=0.25) compared to controls  corresponding to an OR for *6A carriers of 1.3 (95% CI: 1.0–1.9). The OR for *6A/*9A and *6A/*6A genotypes was 1.3 (95% CI: 0.9–1.8) and 2.1 (95% CI: 0.8–6.0), respectively, in this group. There was no difference in the TGFBR1*6A frequency in sporadic or high-risk familial breast cancer compared to controls  for carriers of the 6A allele, this figure based on 1400 breast cancer cases from seven different studies, including the Baxter study , from the population-based cohort, is small and the association of borderline significance only. Moreover, this result needs to be confirmed in a larger material.There are several explanations for difficulties reaching conclusive results on variants in candidate low-penetrance genes such as TGFRB1al cases . We therStudies on Int7G24A are rare. This variant has been reported to be associated with NSCLC, kidney and bladder cancer susceptibility  . The twoTGFBR1 variants *6A and Int7G24A and breast cancer in two breast cancer cohorts from the Stockholm region. The *6A variant seems to confer an increased risk of low-risk familial breast cancer. The Int7G24A variant may be associated with the development of low-risk familial and sporadic breast cancer; however, this needs to be confirmed in a larger material. Furthermore, the *6A variant might be linked to poorly differentiated breast cancer.In conclusion, there was no overall association of the


MaeIII Restriction Fragment Length Polymorphism in exon 3 of the alcohol dehydrogenase II was assessed in serum from 467 randomly selected German women and 278 women with invasive breast cancer to evaluate the interaction between a polymorphism of the alcohol dehydrogenase II gene, alcohol consumption and risk for breast cancer. In both groups, usual consumption of different alcoholic beverages was asked for using semiquantitative food frequency questionnaires. We used multivariable logistic regression to separately estimate the association between alcohol consumption and alcohol dehydrogenase II polymorphism in the population sample and women with breast cancer. The alcohol dehydrogenase II polymorphism was detected in 14 women from the population sample (3.0%) and in 27 women with invasive breast cancer (9.7%). Frequency of alcohol consumption was independent of the genotype in the population sample. In women with breast cancer, there was a significant inverse association between the alcohol dehydrogenase II polymorphism and frequency of alcohol consumption . We observed a gene-environment interaction between the alcohol dehydrogenase II polymorphism, alcohol consumption, and risk for breast cancer. Breast cancer risk associated with alcohol consumption may vary according to the alcohol dehydrogenase II polymorphism, probably due to differences in alcohol metabolism.British Journal of Cancer (2002) 87, 519–523. doi:10.1038/sj.bjc.6600500www.bjcancer.comCancer Research UK© 2002  Alcohol might very well be one of the few modifiable risk factors for breast cancer . DespiteIn humans, alcohol is almost entirely metabolised by oxidative detoxification, mainly dependent on two enzymes, class I alcohol dehydrogenase (ADH) and acetaldehyde dehydrogenase (ALDH). Both enzymes exhibit genetic polymorphisms influencing the rate of conversion from alcohol to acetaldehyde and to acetate . The claa priori hypothesis of an interaction between the ADH2-2 polymorphism alone, alcohol consumption, and the risk for breast cancer in Caucasian women.The aim of the study was to test the We used the case-only design  to evaluThe population sample consisted of women selected from a national health and nutrition survey conducted among healthy people in the western part of Germany in 1987/1988. The survey methodology has been previously described in detail  according to the same protocol in a central laboratory at the Department of Obstetrics and Gynecology of the University of Ulm. The analysis of In both the population sample and women with breast cancer, consumption of different alcoholic beverages was assessed using a semiquantitative food–frequency questionnaire. Women from the population sample were asked about the usual number of days per week on which they consumed different beverages as well as the usual quantity of these beverages consumed per week , smoking , age at menarche (continuous), parity , and history of breast cancer in the mother. The case-only analyses were additionally controlled for menopausal status, age at menopause (continuous), and cumulative years of hormone replacement therapy (continuous)  and their 95% confidence intervals (CI) for the association between categories of alcohol consumption and the ADH2 polymorphism were estimated separately in the population sample and women with breast cancer using multivariable logistic regression. All analyses were controlled for age (continuous), body mass index (weight in kgtinuous) .The mean age of the population sample of 467 women was 54.9 years, slightly lower than the mean age of the 278 cases with invasive breast cancer  and had T1 (43.8%) or T2 (45.2%) tumours. Most cancers were ductal (74.1%) and oestrogen (70.1%) as well as progesteron (65.1%) receptor positive, and 42.1% had already metastasised to lymph nodes at the time of diagnosis (data not shown).In the population sample, 14 heterozygotes of ADH2 (β1β2) were detected (3.0%) compared to 27 heterozygotes in women with breast cancer .The proportion of women consuming alcoholic beverages less often than once a week or more often than once a week was approximately equal (around 40% each) in women from the population sample see . The preP=0.02).Since the frequency of alcohol consumption was independent of the ADH2 polymorphism in the population sample, the possible interaction between the ADH2 polymorphism and frequency of alcohol consumption as to the risk for breast cancer could be assessed using the case-only design. During the last year preceeding the onset of symptoms, 117 women with breast cancer consumed alcoholic beverages less often than once a week and 98 women reported consumption of alcoholic beverages more often than once a week. The prevalence of the ADH2 polymorphism monotonically decreased with increasing frequency of alcohol consumption from 15.4% in women consuming alcoholic beverages less often than once a week to 4.1% in women consuming alcoholic beverages more often than once a week. The case-only analysis confirmed the inverse interaction between the ADH2 polymorphism and frequency of alcohol consumption with respect to risk for breast cancer. The adjusted case-only OR for once a week consumption of alcoholic beverages compared to women consuming alcoholic beverages less often was 0.5 (95% CI: 0.2–1.7). In women consuming alcoholic beverages more often than once a week, this OR was even more pronounced . There was a statistically significant trend for a lower prevalence of the ADH2 polymorphism with increasing frequency of alcohol consumption (−1) of alcohol consumed instead of frequency of consumption, and even more pronounced  in women with oestrogen receptor positive tumours (data not shown).Additional case-only analyses revealed that the observed interaction was independent of menopausal status and any of the known risk factors for breast cancer presented in In a population based case-only study of women with breast cancer we observed a previously unknown interaction between frequency of alcohol consumption, the ADH2 polymorphism, and risk for breast cancer.The use of the case-only study design is dependent on the assumption that the frequency of alcohol consumption does not vary according to the genotype in the population. Since we, like others .in vitro (9.2 min−1 for β1β1 and 400 min−1 for β2β2 isoenzymes)  than Caucasians (0–10%) (enzymes) . FurtherBlood alcohol levels have been associated with oestradiol levels, which play a major role in the development of breast cancer . AlthougP=0.4, data not shown) and changes in the prevalence of the polymorphism over such a short time period are unrealistic. Population stratification or confounding by different genotypes, which are culturally linked to alcohol consumption, is unlikely to introduce relevant bias in this exclusively German population with very similar genetic background (The difference in prevalence of the ADH2 polymorphism in the population sample and in women with breast cancer is intruiging. Is the ADH2 polymorphism a previously unknown risk factor for breast cancer? This difference needs to be interpreted with caution. Women from the population sample were recruited nationwide and approximately 10 years before the recruitment of women with breast cancer. The population sample has been found to be representative of the German population (trans-retinol and 9-cis-retinol, and might play a role in hydroxysteroid metabolism (ADH has been found in different organs not directly involved in alcohol metabolism, including the breast (tabolism . TherefoAlthough we carefully controlled for potential confounding factors in our analysis, the potential for residual confounding by additional factors cannot entirely be ruled out. In particular, the observed risk and interaction may not be entirely due to the ADH2 polymorphism itself, but also to the linkage of ADH2 with other genes that are associated with an increased risk for breast cancer. ADH2 has been observed to be linked to ADH3 (In summary, we observed a previously unknown interaction between alcohol consumption and the ADH2 polymorphism with respect to risk for breast cancer. If replicated, the observed gene-environment interaction might be a good example for the future need of a concept for tailor-made preventive recommendations taking individual genetic susceptibilities into account. It might, however, also be a good example for the complexity and difficulties in applying such recommendations in practice due to the low prevalence of the ADH2 polymorphism in the caucasian population.


Choline and betaine, similar to folate, are nutrients involved in one-carbon metabolism and hypothesised to reduce breast cancer risk. No prospective study among post-menopausal women has examined choline and betaine intakes in relation to breast cancer risk.We examined the intake of choline and betaine and breast cancer risk among 74 584 post-menopausal women in the Nurses’ Health Study. Nutrient intake was assessed using a validated food-frequency questionnaire six times since 1984. During 20 years of follow-up from 1984 until 2004, we documented 3990 incident cases of invasive breast cancer.P-value, test for trend=0.14) for choline and 0.98  for betaine, compared with those in the lowest quintiles of intakes. The results were similar in breast cancer stratified by hormone receptor (oestrogen receptor/progesterone receptor) status. The association between choline intake and breast cancer risk did not differ appreciably by alcohol intake  or several other breast cancer risk factors, including family history of breast cancer, history of benign breast disease, body mass index, post-menopausal hormone use, and folate intake.Overall, choline  and betaine (104±33 mg per day) intake was not associated with a reduced risk of post-menopausal breast cancer. Participants in the highest quintile of intakes had multivariate relative risks of 1.10 : 0.99–1.22; We found no evidence that higher intakes of choline and betaine reduce risk of breast cancer among post-menopausal women. Choline and betaine are nutrients involved in one-carbon metabolism, a network of biochemical reactions that transfer methyl groups from one compound to another . One-carIn 1976, we enrolled 121 700 female registered nurses aged 30–55 years in the Nurses’ Health Study (NHS) . BienniaThe procedures and protocols of the study were approved by the Institutional Review Boards of the Brigham and Women's Hospital.A semi-quantitative food-frequency questionnaire (FFQ) with ∼60 food items was sent to the members of the cohort in 1980. An expanded FFQ with ∼130 food items was administered to women in 1984, 1986, 1990, 1994, and 1998. Participants were asked how often, on an average, they consumed each type of food during the past year. Serving sizes were specified for each food in the FFQ. The questionnaire had nine possible responses, ranging from never or less than once per month to six or more times per day.The choline and betaine composition of individual foods was added to the nutrient database of the FFQ (Harvard University Food Composition Database) using values published by As dietary intake may affect breast carcinogenesis over an extended period of time, to best represent long-term intake, we calculated cumulative averaged intakes of choline and betaine using repeated FFQ data for our primary analysis, to reduce measurement error in dietary assessment (μmol l−1 (P for trend <0.0001), respectively. For the lowest and highest quintiles of betaine intake, the corresponding geometric means were 10.4 and 10.1 μmol l−1 (P for trend 0.05), respectively. The inverse association between dietary choline and betaine intake and levels of blood homocysteine was manifested primarily in participants with low folate intake (P for interaction <0.0001); among those with a folate intake of ⩽250 μg per day, the geometric mean homocysteine concentrations in the lowest and highest quintiles of choline plus betaine intake were 12.4 and 10.2 μmol l−1 (P for trend <0.0001), respectively. Although slightly weaker, a similar inverse association as in the Framingham Offspring Study was found between choline plus betaine intake and plasma homocysteine levels in the NHS. Compared with those in the lowest quintile, individuals in the highest quintile of choline plus betaine intake had 8% lower levels of homocysteine  for confirmation of the diagnosis and for permission to seek relevant hospital records and pathology reports. Only invasive breast cancer cases were included in the analysis. Confirmation of oestrogen receptor (ER) and progesterone receptor (PR) status was obtained by reviewing pathology reports.Participants contributed person-time from the date of return of the 1984 questionnaire until the date of breast cancer diagnosis, death, or June 2004, whichever came first. Participants were divided into quintiles according to their nutrient intakes. Relative risks (RRs) of breast cancer were calculated as the incidence rate for a given quintile of nutrient as compared with the rate for the lowest quintile. We used Cox proportional hazards regression to account for potential effects of other risk factors for breast cancer. To control as finely as possible for confounding by age, calendar time, and any possible two-way interactions between these two time scales, we stratified the analysis jointly on the basis of age in months at the start of follow-up and calendar year of the current questionnaire cycle. Multivariate models also adjusted simultaneously for smoking status, body mass index (BMI) at age 18 years, weight change between age 18 years and current cycle, height, physical activity, age at menarche, family history of breast cancer, history of benign breast disease, parity and age at first birth, post-menopausal hormone use, and intakes of calories, total folate, and alcohol. All covariates except BMI at age 18 years, height, age at menarche, and family history of breast cancer were updated in each questionnaire cycle or whenever new data were available.We also examined the latency effect of choline and betaine intake using multiple dietary assessments. For example, for a latency period of 0–4 years, we used the 1984 intake for cases diagnosed from 1984 to 1986, the 1986 intake for cases diagnosed from 1986 to 1990, the 1990 intake for cases diagnosed from 1990 to 1994, and so on. For latency of 4–8 years, we used the 1984 intake for cases diagnosed from 1988 to 1990, the 1986 intake for cases diagnosed from 1990 to 1994, and so on. The 0–4-year latency analysis corresponds to the analysis using the most recent dietary intake. Latency analyses were used to examine whether there is a possible time lag between choline and betaine intake and risk of breast cancer.P-value for the tests for interaction was obtained from a likelihood ratio test with two degrees of freedom. All P-values were two-sided.The SAS PROC PHREG  tool was used for all analysis, and the Anderson–Gill data structure was used to handle time-varying covariates efficiently . For all RRs, 95% confidence intervals (CIs) were calculated. Tests for trend were conducted using the median value for each category of nutrient as a continuous variable. To test whether the association between choline intake and breast cancer risk was modified by alcohol intake, folate intake, BMI, post-menopausal hormone use, history of benign breast disease, and family history of breast cancer, cross-product terms for the level of an interaction variable  and choline intake expressed as a continuous variable were included in the multivariate model. The P<0.05). Women with higher intake of betaine were less likely to be smokers and more likely to have a history of benign breast disease, lower BMI, and higher folate intake (P<0.05). Choline intake in the NHS has reduced over time; the mean±s.d. values for choline intake were 338±82 mg per day in dietary assessment in 1984 and 298±69 mg per day in 2002, respectively. Betaine intake remained similar over time; the mean±s.d. values for betaine intake were 106±44 mg per day in dietary assessment in 1984 and 100±43 mg per day in 2002, respectively.During 532 338 person-years of follow-up of 74 584 post-menopausal women, we documented 3990 cases of invasive breast carcinoma. The age range of cases at the time of diagnosis of breast cancer was 38–82 years . P-value test for trend=0.14). Betaine intake was not associated with breast cancer risk. None of the individual sources of choline was related to breast cancer risk. Intake of methionine, another nutrient involved in one-carbon metabolism, was not associated with risk of breast cancer; the RR for top vs bottom quintiles of methionine intake was 1.08 (0.97–1.20). Intakes of choline and betaine were not associated with risk of breast cancer in situ .Higher intakes of dietary choline and betaine were not associated with reduced breast cancer risk . Cholinevs bottom quintiles of choline intake were 1.07 (0.92–1.25) for ER+/PR+ cancers (n=1998), and 1.15 (0.86–1.54) for ER−/PR− cancers (n=538). As there may be a latency effect of choline and betaine, we took advantage of repeated assessment of dietary intake to assess the temporal relationship between choline and betaine intake and breast cancer risk. We evaluated latency periods of 0–4, 4–8, 8–12, 12–16, and 16–20 years. When the latency period became longer, the modest positive association between choline intake and breast cancer became weaker and in fact became slightly inverse for a latency period of 16–20 years (RR for the top vs bottom quintile of choline intake=0.92 (0.74–1.14)). There was no appreciable difference in association between betaine intake and breast cancer risk by latency period.Results were similar when breast cancers were grouped by hormone receptor status. For example, the RRs for top μg per day) and alcohol intake  . The association between choline intake and breast cancer risk did not differ appreciably by alcohol intake .The availability of other dietary factors related to one-carbon metabolism, including alcohol and folate, may modify the association between choline and breast cancer risk. We thus examined choline intake and breast cancer risk by levels of folate intake  . The pos−2), and post-menopausal hormone use  ; P for trend 0.03), but not among those with a history of benign breast disease (RR for the top vs bottom quintile 0.98 (95% CI: 0.84–1.13); P for trend 0.98). Similarly, choline intake was positively associated with breast cancer among current users of post-menopausal hormone (RR for the top vs bottom quintile 1.32 (95% CI: 1.03–1.70); P for trend 0.06), but not among those who never used hormone or who used hormone in the past (RR for the top vs bottom quintile 1.12 (95% CI: 0.94–1.33); P for trend 0.17 for never users and RR for the top vs bottom quintile 1.03 (95% CI: 0.88–1.21); P for trend 0.94 for past users).We also examined whether the association between intakes of choline and betaine and breast cancer risk differed by levels of other breast cancer risk factors, such as family history of breast cancer , history of benign breast disease , BMI  . None ofIn this large prospective study among post-menopausal women, we did not find any strong overall associations between intakes of choline and betaine and risk of breast cancer. The results were similar in breast cancer stratified by hormone receptor status and in the analyses stratified by several breast cancer risk factors.P-value for trend=0.29) (Our findings on choline intake are not consistent with a case–control study of 1508 breast cancer cases that found that higher choline intake was associated with up to 24% lower risk of breast cancer, although the association was not linear (de novo synthesis in the body. The recommended daily intake was set in 1998 at 550 mg per day for men and 425 mg per day for women (Choline and betaine are nutrients that were, until recently, not extensively investigated in epidemiological studies because of lack of food composition databases . NutrienThe positive association between choline intake and breast cancer risk, although modest and not significant, is similar to that which we found in relation to colorectal adenomas in this population; we found that higher choline intake was associated with up to 45% elevated risk of colorectal adenomas (The suggestive interaction between choline and folate intake in relation to breast cancer risk is hard to interpret, especially because the positive association between choline intake and breast cancer risk was mostly limited to middle-intake categories of folate. For colorectal adenomas, we found that the positive association between choline intake and adenoma risk was the strongest among those with the lowest folate intake (This study had several strengths. The prospective nature of the study avoided biases of case–control studies. Our study provided a unique opportunity to evaluate intakes of choline and betaine in a large population of post-menopausal women, which provided informative numbers of breast cancer cases in several stratified analyses. As we had repeated measures of dietary intake, we were able to examine long-term averaged diet, as well as several latency analyses. Measurement error is reduced by the use of repeated measures. Finally, we had information on a wide range of potential confounders and adjusted for them.In conclusion, in this cohort of post-menopausal women, we found that intakes of choline and betaine were not related to a reduced risk of breast cancer.


Although a role of childhood body size in postmenopausal breast cancer risk has been established, less is known about its influence on tumour characteristics.P values for heterogeneity were obtained by performing one degree of freedom trend tests.We studied the relationships between childhood body size and tumour characteristics in a Swedish population-based case-control study consisting of 2,818 breast cancer cases and 3,111 controls. Our classification of childhood body size was derived from a nine-level somatotype. Relative risks were estimated by odds ratios with 95% confidence intervals, derived from fitting unconditional logistic regression models. Association between somatotype at age 7 and tumour characteristics were evaluated in a case-only analysis where P trend = 0.004) after adjustment. The significant protective effect was observed within all subgroups defined by estrogen receptor (ER) and progesterone receptor (PR) status, with a stronger effect for ER-negative , than for ER-positive , tumours (P heterogeneity = 0.046). Somatotype at age 7 was not associated with tumour size, histology, grade or the presence or absence of metastatic nodes.A large somatotype at age 7 was found to be associated with decreased postmenopausal breast cancer risk. Although strongly associated with other risk factors such as age of menarche, adult body mass index and mammographic density, somatotype at age 7 remained a significant protective factor (odds ratio (OR) comparing large to lean somatotype at age 7 = 0.73, 95% confidence interval (CI) = 0.58-0.91, Greater body size at age 7 is associated with a decreased risk of postmenopausal breast cancer, and the associated protective effect is stronger for the ER-negative breast cancer subtype than for the ER-positive subtype. There is considerable evidence that childhood anthropometric measurements are associated with postmenopausal breast cancer risk. It has been consistently shown that variables that approximate body shape and size early in life are inversely associated with breast cancer risk in adulthood. For example, a study conducted in 1998 on the same data set as used in the current study  found thAlthough a role of childhood body size in adult breast cancer risk has been established, less is known about its influence on tumour characteristics. One study by Bardia and colleagues  looked iThe subjects included in the current study are drawn from a population-based case-control study of postmenopausal breast cancer in Swedish-born women aged 50 to 74 years at the time of enrolment, which was between 1 October, 1993 and 31 March, 1995. Controls were randomly selected from the Swedish registry and frequency matched to the expected age distribution of the cases. Details on data collection and subjects have been described previously [2), history of hormone replacement therapy (HRT) , and family history of breast cancer . Age at menopause  was also derived from information collected in the study questionnaire and the definition used in this study has been previously described [With the exception of clinical data on tumour characteristics and mammographic density, all other covariate data were derived from the parent case-control study. Anthropometric measurements at age seven years and one year prior to enrolment were collected by means of a nine-level somatotype Figure  featuredescribed . It is dInformation regarding the retrieval of tumour characteristics from the medical records of all participants from surgical and oncological units throughout Sweden have been presented in detail elsewhere ,10. The The process of collecting mammographic density data in this study has been described previously . Film maThe distribution of baseline characteristics of known breast cancer risk factors were summarised as means and standard deviations or proportions. Odds ratio (OR) estimates with corresponding 95% confidence intervals (CI) were computed by fitting unconditional logistic regression models with breast cancer risk status as the response variable, adjusting for age.a priori evidence in the published literature of the factor being associated with both childhood body size and breast cancer risk, or if the factor was significantly associated at the 5% level with both somatotype at age seven years and breast cancer risk. Those covariates that, when added to the model, changed the coefficient by more than 10%, were considered confounders and adjusted for in the multivariate analysis. The final variables in the multivariate logistic regression model examining breast cancer risk overall, and stratified by ER and PR tumour subtypes, included age, age at menarche, benign breast disease, and BMI one year prior to enrolment (recent BMI). Adjustment for other variables did not influence the somatotype risk estimates. Mammographic density was also identified as a confounder. However, as mammographic density data are only available for a subset of the subjects, this variable was accounted for together with the other risk factors in a separate model. Women with and without mammographic density information were not found to differ significantly at the 5% level for the covariates included in the analysis models (data not shown).To identify potential confounders of the association between somatotype at age seven years and breast cancer risk, linear/logistic regression models were fitted for either continuous  or binary (benign breast disease and HRT) outcomes including only controls in the analysis. Somatotype at age seven years was treated as a categorical (three-level) independent variable. Proportional odds logistic regression was used in situations where the outcome variable was ordinal (somatotypes at age seven years and one year prior to enrolment) from which cumulative OR esimates with corresponding 95% CIs were computed. Covariates were considered potential confounders if there was P values for heterogeneity were obtained by performing one degree of freedom trend tests. As there exists prior evidence that certain tumour characteristics such as ER status are associated with age at diagnosis [Associations between somatotype at age seven years and tumour characteristics were evaluated in a case-only analysis, by fitting ordinal regression models treating tumour characteristics as dependent variables, with somatotype at age seven years included as a covariate. iagnosis , and thaiagnosis . The levP = 0.057), a result consistent with the literature [P = 0.001). A larger proportion of cases than controls had a leaner body shape at age seven years. Despite somatotype one year prior to enrolment having a high correlation to recent BMI , it was not found to be significantly associated with breast cancer .Table S1 in Additional file terature . Family P trend < 0.0001), benign breast disease , and mammographic density , somatotype one year prior to enrolment  and recent BMI . No evidence of association was found between age of menopause and somatotype at age seven years or between family history and somatotype at age seven years. Parity and HRT were found to be independent of somatotype at age seven years .To identify potential confounders of the association between somatotype at age seven years and breast cancer risk, we assessed whether other established risk factors for breast cancer are associated with somatotype at age seven years. An increase in childhood body size was found to exhibit strong inverse associations with age of menarche . The protective effect of a larger somatotype was found to be significant (P trend < 0.05) for ER-negative, PR-positive and PR-negative subtypes and marginally significant (P trend = 0.062) for the ER-positive subtype. Within the group consisting of large somatotypes, the most prominent effects were shown in ER-negative  and PR-negative  tumours. The point estimates changed very little before and after additional adjustment for mammographic density as a continuous variable [see Table S2 in Additional file After adjustment of known breast cancer predictors and other associated risk factors, the inverse association of somatotype at age seven years with breast cancer remained highly significant . The point estimates for tumour size, histology, grade, or the presence or absence of metastatic nodes did not vary much before and after adjustment for age of diagnosis as a continuous variable.We next assessed the effects of childhood body size on tumour characteristics  by fitting binary/ordinal logistic regression models, adjusting for age at diagnosis in years as a confounder. We established that the protective effect of somatotype at age seven years was significantly stronger for ER-negative disease than for ER-positive disease (P trend = 0.004) after adjustment for these other risk factors. Our second and most novel finding was of a significant protective effect of somatotype at age seven years regardless of receptor status, but with a stronger effect for ER-negative , than for ER-positive , tumours (P heterogeneity = 0.046).Our first main finding was that a large somatotype at age seven years was associated with a decreased risk of postmenopausal breast cancer. Although strongly associated with other risk factors such as age of menarche, adult BMI and mammographic density, somatotype at age seven years remained a significant protective factor , but not the PR subtypes. However, in lieu of the fact that PR is an estrogen-induced target gene, and that its presence could serve to indicate ER functional capacity and tumour differentiation state [The age-adjusted case-only comparison of our data reflected a significant difference in the effects of childhood body size on the two ER subtypes , thus suggesting an independent underlying mechanism. We speculate that a possible mechanism driving the negative association with breast cancer risk could be epigenetic changes that occur during mammary development. Hilakivi-Clarke [Hormonal exposure and mammographic density are established risk factors of breast cancer that have been suggested to be independent, operating through different pathways . Adjustmi-Clarke  summarisi-Clarke . Prepubei-Clarke . Liu andi-Clarke  also demi-Clarke . The eviThe strengths of our study include it being a population-based study, its large sample size and detailed information on many variables: anthropometric measures at different time points throughout life, mammographic density, reproductive and hormonal risk factors, and tumour characteristics. To our knowledge, this is the first study to consider the effects of somatotype at age seven years on adult breast cancer with the consideration of mammographic density, and also the first to examine its effects on tumour characteristics other than ER status.A limitation of our study is that risk factor data were self-reported, and could thus be measured with error. Although two studies have demonstrated the validity of using the nine-level somatotype diagram for the long-term recall of childhood body size via high correlations with BMI at the same ages ,36, it iOur findings may have important implications. The effects of childhood body size on the different breast cancer subtypes are independent of other breast cancer risk factors, such as mammographic density and estrogen exposure. Given the strength of the associations, and the ease of retrieval of information on childhood somatotypes retrospectively from pictures early in life, childhood body size is potentially useful for building breast cancer risk or prognosis prediction models. It appears counterintuitive that a large body size during childhood can reduce breast cancer risk or alter one's prognosis, because a large birth weight and a high adult BMI have been shown to otherwise elevate breast cancer risk. There remain unanswered questions on mechanisms driving this protective effect. Because body size and related hormonal exposures are modifiable risk factors, women might substantially decrease their risk of breast cancer, in particular the more aggressive ER-negative disease, by monitoring their nutrition and exogenous hormone intake at different points in life.BMI: body mass index; CI: confidence interval; ER: estrogen receptor; HRT: hormone replacement therapy; OR: odds ratio; PR: progesterone receptor.The authors declare that they have no competing interests.JLi participated in the study design, carried out the analyses and drafted the manuscript. LE digitised and obtained readings for the mammograms. KH, KC, JLiu and PH participated in study design and coordination and helped to draft the manuscript. All authors read and approved the final manuscript.Table S1. Descriptive characteristics of post-menopausal women.Click here for fileTable S2. Multivariate-adjusted odds ratio (OR) estimates and corresponding 95% confidence intervals (CIs) of postmenopausal breast cancer for somatotype at age seven years on a subset of women with mammographic density data; overall and stratified by breast cancer tumour subtype based on estrogen receptor (ER) and progesterone receptor (PR) status.Click here for file


The suprachiasmatic nuclei in the hypothalamus, one of the most important physiological determinants of alertness and performance, drive a circadian pacemaker in mammals, with an intrinsic period averaging 24 h. Light is the primary stimulus to the disruption and resetting of this pacemaker, which is expressed in changing melatonin rhythms. Melatonin production in humans decreases when people are exposed to light at night. Since melatonin shows potential oncostatic action in a variety of tumours, it is possible that lowered serum melatonin levels caused by exposure to light at night enhance the general tumour development. Cancer is the second leading cause of death in industrialised countries like the United States, where a significant proportion of workers engage in shift work, making a hypothesised relation between light exposure at night and cancer risk relevant. Observational studies support an association between night work and cancer risk. We hypothesise that the potential primary culprit for this observed association is the lack of melatonin, a cancer-protective agent whose production is severely diminished in people exposed to light at night. Light at night is an environmental exposure that has become integral to our lives. Before 1879, when the light-bulb was invented, humans were exposed to only insignificant amounts of low-intensity light at night from sources such as candles and petroleum lamps. Later on, paralleled by growing industrialisation, a need for longer days emerged, with electric light making it possible to push multifaceted working schedules further and further into the night. Today, exposure to light at night, both in the form of occupational exposure during night work and as a personal choice and lifestyle, is experienced by numerous night-active members of our societies. We, however, do not appear to be well adapted to bright nights: shift workers, for instance, reportedly suffer from a variety of health problems . PreviouPast studies suggested that the excess breast cancer mortality seen in urban areas as well as the North East as compared to the South of the US could be due to differences in exposure to sunlight .P=0.02). Among postmenopausal women, the relative risk for breast cancer, controlling for all the major risk factors for breast cancer, was moderately increased after 1–14 and 15–29 years of rotating night shift work, and was further increased  for those nurses who worked the night shift for 30 or more years, with similar risks for premenopausal women . Thus, in sum, observational studies seem to support the hypothesis that night work increases the risk for breast cancer. This association may be mediated, at least in part, by the oestrogen pathway.Observational studies have supported that theory, indicating that women in occupations that expose them to light at night do experience a higher risk of breast cancer, and that blind women, who do not have the ability to experience lower melatonin levels because of their supposed lack of receptivity to light, have a lower incidence of breast cancer . Studieset al report an increased risk of colon  and rectum cancer  in their cohort of female radio and telegraph workers. Rafnsson and colleagues do not report the risks for colorectal cancer among the female Icelandic flight attendants, but describe an elevated risk for tumours of the lymphatic system.Only few previous observational studies have addressed the relationship between shift work and cancers, other than breast cancer. Early suggestions for an increased cancer risk related to shift work arose from two mortality studies that were conducted among male shift workers to assess the influence of shift work upon total and cause-specific mortality, with suggestions for an increased cancer mortality related to shift work. Tynes In vitro studies, although not entirely consistent .Given the evidence from experimental studies supporting the general oncostatic property of melatonin, we therefore speculate that exposure to light at night not only has an impact on breast cancer risk, but also may increase the risk of other cancers, primarily through the melatonin pathway. This has been posed previously (On the basis of the observed effect of night work on colorectal cancer, which, contrary to breast cancer cannot be fully explained by variations in circulating oestrogen levels, taken together with consistent evidence from experimental studies, we believe that the effects of light exposure at night on cancer risk may be explained, at least in part, by the direct oncostatic mechanisms of melatonin. We propose that an increase in light exposure during the night decreases the amount of time available for melatonin production, which reduces the possible nonspecific oncostatic effect of the pineal gland, thus increasing the risk not only of breast cancer but also of other cancers. Although other aetiologic mechanisms, such as the loss of normal diurnal variation in cortisol , may be


Height was positively associated with risk in both cohorts . Body mass index at the age of 18 years and birthweight were not associated with risk. Further research should examine the biological mechanisms underlying the observed associations.Adult body mass index (BMI) has been associated with ovarian cancer risk, but few studies have examined body size earlier in life. We prospectively examined associations of body fatness at ages 5 and 10, BMI at age 18, height, and birthweight with risk of epithelial ovarian cancer in the Nurses' Health Study  and Nurses' Health Study II . Cox proportional hazards regression was used to estimate relative risks (RRs) and 95% confidence intervals (CIs). There was a weak inverse association between average body fatness at ages 5 and 10 and risk in the NHS (RR for heaviest  Overweight and obesity have been associated with risk of cancer in women , but theTherefore, we prospectively examined the associations of body size in early life – including body fatness at ages 5 and 10 years, BMI at the age of 18 years, birthweight, and height – with risk of epithelial ovarian cancer among participants in two large cohort studies, the Nurses' Health Study (NHS) and the NHSII.The NHS began in 1976 and the NHSII in 1989, when 121 700 and 116 609 US female registered nurses, respectively, completed a mailed questionnaire about their lifestyle factors, health behaviours, and medical histories. Follow-up questionnaires have been sent to participants every 2 years since enrolment. Incident cases of epithelial ovarian cancer were reported on the biennial questionnaires through 2004 (NHS) and 2005 (NHSII). A gynaecologic pathologist reviewed the pathology reports and medical records to confirm the diagnosis and identify histological type, subtype, morphology, and stage  at ages 5 and 10 years using a nine-level figure drawing, where level 1 represents the most lean and level 9 represents the most overweight   were combined in the analysis to increase power. In a NHSII validation study (Age and other covariates were assessed on the questionnaires throughout the study.Participants contributed person-time from baseline (the questionnaire year that the exposure of interest was assessed) until the date of ovarian cancer diagnosis, report of other cancer (except nonmelanoma skin cancer), death, or 31 May 2004 (NHS) or 31 May 2005 (NHSII), whichever occurred sooner. We excluded women reporting a previous diagnosis of cancer except nonmelanoma skin cancer and those with a history of bilateral oophorectomy or pelvic irradiation. For analyses focusing on each individual body size measure, we excluded women who were missing data for that measure. The years of assessment, numbers of eligible participants and cases, and total numbers of person-years available for each analysis are shown in Cox proportional hazards models stratified by age in months and 2-year questionnaire cycle were used to estimate relative risks (RRs) and 95% confidence intervals (CIs), adjusting for ovarian cancer risk factors. Tests for linear trend were conducted by including each body size measure in a model either as an ordinal variable  or as a continuous variable with values equal to the category medians .We first conducted the analyses within the NHS and NHSII separately. We then evaluated heterogeneity in the estimates by cohort . We usedMost adult characteristics were not associated with early life body size . In bothvs level 1=0.69, 95% CI: 0.48–0.99, P for trend=0.01). Averaging ages 5 and 10 years, the RR for childhood body fatness level ⩾5 compared to level 1 was 0.81 . In contrast, there was some suggestion of a positive association for average childhood body fatness in the NHSII , although this was not statistically significant (P for trend=0.10) . Body mass index at the age of 18 years was not significantly associated with risk in either cohort.In the NHS, greater body fatness at ages 5 and 10 years were associated with decreased risk of ovarian cancer , althougnd=0.10) . The assvs level 1=1.38, 95% CI: 0.70–2.71, P for trend=0.92) and a weak inverse association in postmenopausal women , although neither these nor the interaction with menopausal status (P=0.37) were statistically significant. Results for BMI at the age of 18 years were similar in premenopausal women, but there was no evidence of an inverse association in postmenopausal women (P for interaction=0.11).One major difference between the NHS and NHSII cohorts is the menopausal status of participants when childhood body size was assessed . To explore whether this could explain the observed difference in the association of childhood body fatness with risk, we combined the data from both cohorts and stratified by menopausal status in the cycle before diagnosis . There wvs level 1=1.77, 95% CI: 0.85–3.69, P for trend=0.39) and in women between the ages 50 and 59 years , whereas the association was inverse in women at the age of 60 years and older . The interaction between childhood body fatness and age was statistically significant (P=0.001). When we jointly stratified by age and menopausal status, the positive association between childhood body fatness and ovarian cancer risk appeared stronger in premenopausal women under the age of 45 years  than in those at the age of 45 years and older . Conversely, the inverse association in postmenopausal women was stronger among those at the age of 60 years and older  than in those younger than the age of 60 years .Alternatively, the observed variation in the associations for childhood body fatness could be explained by age differences between participants in the two cohorts ; therefore, we conducted a preliminary analysis combining both cohorts and stratifying by age, while adjusting for menopausal status. There were nonsignificant positive associations between childhood body fatness and ovarian cancer risk in women less than the age of 50 years . Adjustment for age at menarche had no substantial impact on the associations (data not shown).Additional adjustment for current BMI as a continuous variable had virtually no impact on the childhood body fatness associations in the NHS or for postmenopausal women overall (data not shown). However, the positive association for childhood body fatness in the NHSII and in premenopausal women was attenuated when including current BMI , and the test for heterogeneity was not significant (P=0.22); however, the association appeared stronger in the NHSII  than in the NHS  (Height was positively associated with ovarian cancer risk in both cohorts (pooled RR for ⩾1.75 nd=0.01) . The posThe observed associations for body size in early life were similar for invasive cases alone and by histological type, although these analyses were limited by small case numbers. No significant interactions were observed between any of the body size measures and parity, oral contraceptive use, postmenopausal hormone use, or family history of breast or ovarian cancer (data not shown).Results from this study indicate that body size in early life may be related to the risk of epithelial ovarian cancer. Body fatness during childhood was associated with ovarian cancer risk, although the association differed by cohort; greater body fatness at ages 5 and 10 years was associated with a slightly lower risk among NHS women, but a suggestion of an increased risk among NHSII women. These differences could be explained by differences in the menopausal status or age of participants. Body mass index at the age of 18 years and birthweight were not associated with risk; however, height was positively associated with risk.To our knowledge, this is the first study to investigate the association of childhood fatness with ovarian cancer risk. Previous studies have examined the relation between overweight and obesity in adulthood and ovarian cancer risk or mortality, with mixed results (A second potential explanation is that the relation between body size at young ages and ovarian cancer risk may differ by menopausal status or age. In a pooled analysis of 12 cohorts, BMI at baseline was not associated with risk overall or among postmenopausal women, but there was a positive association in premenopausal women (However, our findings of different associations for childhood body fatness with ovarian cancer risk by cohort could be partially due to a cohort effect. Greater childhood body fatness among women in the NHS may have been determined by different factors than in the NHSII. Chance could also explain our results, given that the observed associations for childhood body fatness were modest and the trends were only marginally significant. However, strong associations between childhood body fatness and risk of breast cancer have been observed in previous studies, suggesting that it may predict the risk of hormone-related cancers in women. Further, previous studies have observed significant associations of both adolescent and adult BMI with ovarian cancer risk, lending plausibility to our findings.The associations of body size in childhood with ovarian cancer risk could be mediated through endogenous hormones , althougThe observed positive association between height and ovarian cancer risk is consistent with results from previous studies in several populations . InteresDespite consistency across epidemiological studies, the biological mechanisms explaining the observed associations for height are unclear. Adult height may be a marker for genetic factors or of energy intake, caloric restriction, or exposure to sex and growth hormones in early life . Howeverin utero and during infancy, and that this could influence ovarian cancer pathogenesis later in life. A population-based case–control study found no overall association between birthweight and ovarian cancer risk, although among women younger than the age of 55 years, there was a decreased risk of ovarian cancer for those who weighed <5.5 pounds at birth compared to those who weighed 5.5–9 pounds (There are almost no epidemiological data on the relation of birthweight to risk of ovarian cancer. In a small retrospective medical record review, birthweight was not associated with mortality from ovarian cancer, but greater weight gain in the first year of life was associated with increased mortality (Our study has several limitations. Although we combined data from two large cohort studies, we had limited power to examine interactions or variation by histological type. Another limitation is the reliance on recall of body size in early life. Important strengths of our study include its large sample size, its prospective design, the confirmation of ovarian cancer cases, and the detailed information on menopausal status and ovarian cancer risk factors.In summary, this study suggests that body fatness during childhood and adult height may be related to ovarian cancer risk, and that these associations may vary by menopausal status or age. This is the first study to examine the associations of body fatness during childhood with ovarian cancer risk, and it is one of the largest, most comprehensive studies of the other body size measures to date. Further research should confirm these findings in other populations as well as examine the underlying biological mechanisms.


Leu33Pro heterozygotes, homozygotes and heterozygotes+homozygotes vs noncarriers had odds ratios for breast cancer of 1.0 , 0.8 (0.5–1.2) and 1.0 (0.8–1.1), respectively. After stratification for conventional risk factors, odds ratio for breast cancer in heterozygotes, homozygotes and heterozygotes+homozygotes vs noncarriers were not increased above 1.0 in any of the 14 strata examined. This was also true after stratification for tumour histological subtype and cancer stage at the time of diagnosis.To pursue a borderline increased risk of breast cancer for carriers of two integrin beta Third, ITGB3 Leu33Pro heterozygosity and homozygosity have an increased risk of invasive breast cancer. Given the prospective data, the assumption of a dominant genetic effect seems reasonable, so we also combined the heterozygous and the homozygous participants in one group vs noncarrier participants. We recruited 1088 consecutive female patients with invasive breast cancer and compared these with 4815 women from the general population without breast cancer. This large case–control study also allowed us to search for genotype-associated increased risk of breast cancer in women stratified for conventional risk factors, and to determine whether genotype-associated breast cancer risk is restricted to certain histological subtypes or cancer stages.For this reason, we tested the hypothesis that n=1088) were consecutively collected from the Department of Breast Surgery, Herlev University Hospital from 2001 to 2004. Owing to centralised breast cancer care, this department covers all cases of invasive breast cancer in roughly half of the Greater Copenhagen area, a population amounting to 320 000 women the 1st of January 2004. All patients filled in a questionnaire on lifestyle and health, including reproductive, menstrual and hormonal history, alcohol consumption, height and weight. Premenopausal women were defined as women who had a menstrual period within the previous 2 months. Postmenopausal women had a period more than 12 months ago. If the menstrual period was 2–12 months, the level of follicular-stimulating hormone decided whether the patient was pre- or postmenopausal. Furthermore, women >55 years of age after hysterectomy or with menstruation on hormonal substitution were considered postmenopausal. One or more questionnaire informations were missing for 546 women. However, for test of the main hypothesis (genotype association with risk of breast cancer adjusted for age), full information was available on all 1088 breast cancer cases.Cases of invasive breast cancer in women  of the Copenhagen City Heart Study , and the scientific ethical committee of the County of Copenhagen approved the collection of cases (KA 02152). The studies were also approved by Herlev University Hospital.3 integrin gene (GenBank Accession no. NM000212.1), which introduces an MspI recognition site. The assay also included a second MspI recognition site always cleaved, which served as a control site for the digestion reaction. A 268 bp fragment of exon 3 was amplified from genomic DNA with flanking intronic primers, cleaved with MspI, run on a 3% agarose gel and visualised by staining with ethidium bromide. Genotypes were determined independently by an experienced lab technician and an author (SEB).Participants were genotyped as described earlier , use of oral contraceptive pill (yes/no), menopausal status  and use of hormonal replacement therapy (yes/no). In the analysis of histological subtype and invasive breast cancer stage, each subtype/stage was compared to the controls with the age-matched model of 1-year age strata while excluding cases without the analysed subtype/stage.We performed a matched case–control study with strata of 1, 3 and 5 years of age. This resulted in 63 one-year strata with a mean of 4.5 controls per case, 22 three-year strata and 4.5 controls per case and 14 five-year strata and 4.6 controls per case, respectively. Using 1-year strata, two cases both aged 93 were unmatched with controls and 168 controls aged less than 28 were unmatched with cases. Conditional logistic regression was used to calculate odds ratio for invasive breast cancer according to genotype. Two models were applied: an age-matched model and an age-matched model also adjusted multifactorially for body mass index (continuous), alcohol consumption (continuous), parity (0 P=0.67 and 0.70).Characteristics of participants are shown in vs noncarriers were 1.0 , 0.8 (0.5–1.2) and 1.0 (0.8–1.1), respectively , this association became nonsignificant after correcting for multiple comparisons. Furthermore, this association was statistically nonsignificant in the age-matched study with more statistical power.To explore whether genotype interacts with conventional risk factors for breast cancer, we stratified the age-matched and multifactorially adjusted analyses by age, body mass index, alcohol consumption, parity, use of oral contraceptive pills, menopausal status and use of hormonal replacement therapy. Risk of breast cancer in homozygotes, heterozygotes and heterozygous+homozygous examined . Althougvs noncarriers did not differ from 1.0 in any histological subgroup of invasive breast cancer Risk of breast cancer in homozygotes, heterozygotes and heterozygous+homozygous vs non-carriers did not differ from 1.0 in any cancer stage subgroup  and 1.9 (1.0–3.7), P-value 0.06, so this could represent a chance finding. In contrast to another study, vs noncarriers in a case–control study of 100 cases and 100 controls. However, genotype distribution among controls was not in Hardy–Weinberg equilibrium (P=0.048) and the finding was only significant at P=0.04, so this also could represent a chance finding. In a third study (vs noncarriers were 1.1 (0.9–1.4), 1.3 (0.7–2.4) and 1.1 (0.9–1.4), respectively. A borderline (P=0.055) interaction between age (age ⩽45 vs >45 years) and genotype (noncarriers vs heterozygotes+homozygotes) with an odds ratio of heterozygotes+homozygotes vs noncarriers of 1.3 (1.0–1.8) in women aged over 44 years; in our study, no such interaction could be detected (P=0.64). They also detected no association between tumour stage or grade and genotype (noncarrier vs heterozygous+homozygous state) after correcting for multiple comparisons.So far, four papers have been published on this polymorphism and breast cancer risk . Our ownrd study  of 602 cLike the previous studies mentioned, the present study has certain limitations. Covariates for controls were recorded at the 1991–1994 examination, while those for the cases were recorded at the time of diagnosis in 2001–2004, approximately 10 years apart. For conventional risk factors such as the use of hormonal replacement therapy, this could distort risk estimations, but will not invalidate analyses of genetic risk factors, as genotypes are permanent. Although cases and controls were recruited from slightly different geographical locations of the greater Copenhagen area, the background population is the same, more than 97% being white women of Danish descent in both cases and controls.ITGB3 Leu33Pro heterozygosity and homozygosity do not increase the risk of breast cancer.Overall, the combined evidence from this and previous reports indicates that ITGB3 Leu33Pro homozygotes, but should rather try to examine other cancer subgroups like ovarian cancer, melanoma or yet other cancer subgroups (Future research should no longer focus on risk of breast cancer in


In a case–control study of kidney cancer in four central European countries, with 1097 incident cases and 1476 controls, we found an increased risk for self-reported hypertension and for obesity. Additional unknown risk factors are likely to be responsible for the high rates of kidney cancer in this region. Kidney cancer accounts for 1.9% of all malignancies, with approximately 189 000 new cases diagnosed globally each year and the incidence varying more than 10-fold around the world, with the highest rates generally observed in central European countries and among blacks in the United States . Smokingn=1476) were admitted to the same hospitals as the cases for conditions unrelated to smoking or genitourinary disorders (except for benign prostatic hyperplasia) between August 1998 and March 2003. No single disease made up more than 20% of the control group. Both cases and controls had to be residents of the study areas for at least 1 year. The response rate for cases ranged from 90 to 98.6% and for the controls, 90.3–96.1%.This hospital-based case–control study was conducted in seven centres (Moscow (Russia), Bucharest (Romania), Lodz (Poland), and Prague, Olomouc, Ceske Budejovice, and Brno (Czech Republic)). A total of 1097 newly diagnosed, histologically confirmed, renal parenchymal cancers (ICD-O-2 code C64) between 20 and 79 years of age were recruited during August 1999 and January 2003. Trained medical staff reviewed medical records to extract relevant diagnostic information, including date and method of diagnosis, histologic type, tumour location, stage, and grade. Eligible controls  was defined as status 2 years before interview. Packyears were used as a measure of cumulative tobacco smoking and were calculated by the number of cigarettes smoked per day multiplied by years of smoking and divided by 20. History of hypertension was self-reported and a positive history was restricted to patients who reported being treated for hypertension. The weight in kilograms was ascertained for 2 years before the interview. Body mass index was calculated by dividing the weight by the square of the height in metres.vs yes), and country, where appropriate. Heterogeneity tests were used to evaluate differences among countries by including country–exposure interaction terms in the logistic models, and likelihood ratio tests to evaluate the statistical significance of the interaction terms.Kidney cancer risks were estimated by odds ratios (ORs) and 95% confidence intervals (CIs) using logistic regression analysis, with adjustment for age , smoking , BMI in five categories , history of hypertension treatment (no The study population consisted of 1097 kidney cancer cases (648 men and 449 women) and 1476 controls (952 men and 524 women)  and 2. ASmoking was not associated with an increased risk of kidney cancer, and no dose–response was seen with increasing levels of cigarette consumption. The findings were similar for all patients combined or in men and women separately. In addition, no significant heterogeneity between the four countries was observed.P=0.011; Increasing BMI was positively associated with kidney cancer overall ) . The effOur analysis included all histological types of carcinoma of the renal parenchyma, including 17% of cases who had an histology other than clear cell type, excluding these cases had little influence on the results.This study provides evidence that increased BMI and a history of hypertension are risk factors for kidney cancer in high-risk areas of central and eastern Europe, countries in which data on kidney cancer have been limited. The increased risk associated with BMI was more pronounced among men, while the increased risk for history of hypertension was more pronounced among women.Our findings of increasing risks for kidney cancer with increasing BMI agree with most other studies. A recent evaluation by an IARC working group on the effects of body weight and adiposity showed that all but one of the 19 reviewed studies found a more than two-fold increase in kidney cancer risk among obese men and women compared with those of normal weight . In addiAn association between hypertension and kidney cancer has also been previously reported . The relSmoking is an established risk factor for kidney cancer, although the increase in risk is weak . In a coOur study confirms that BMI and a history of hypertension are risk factors for kidney cancer in central Europe, a region with the highest incidence currently reported, but these are unlikely to explain fully the high incidence, indicating that important causes remain to be discovered.


Parity has been linked to gallbladder cancer and gallstones, but the effects of other reproductive factors are less clear.We examined 361 incident biliary tract cancer cases, 647 biliary stone cases, and 586 healthy women in a population-based study in Shanghai.vs 1 child⩾3 =2.0, 95% confidence interval (CI) 0.7–5.1), younger age at first birth , and older age at menarche  on gallbladder cancer risk were more pronounced among women with stones, but the interactions were not significant.The effects of parity (odds ratios, OROur results provide support for high parity, younger age at first birth, and late age at menarche in the development of gallbladder cancer, particularly among women with biliary stones. Biliary tract cancers, encompassing tumours of the gallbladder, extrahepatic bile ducts, and ampulla of Vater, are uncommon but highly fatal . The incHigh parity is considered as a risk factor for gallstones . Moreoven=31). In all, 70% of cancer cases were confirmed by histopathology, whereas the remaining were confirmed with medical records. We included 647 female patients with stones  but without cancer; stone cases were confirmed with medical records. A total of 586 healthy women without cancer or stones were randomly selected from the Shanghai population. Stone status was assessed in nearly all cancer cases and controls using self-reported history and medical records.Details of the study have been described previously , bile duct cancers with all controls (n=586), and stones with controls without stones (n=422). Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated using unconditional logistic regression adjusted for age, education, and body mass index. Continuous variables were categorised into quartiles or tertiles on the basis of distributions within controls and those reported among Chinese women . Compared with women who had one child, those having three or more children had a 2.1-fold risk (95% CI 1.0–4.2) of gallbladder cancer. Among parous women (97% of subjects), those who had their first child before 27 years of age had higher risks of gallbladder cancer relative to those who had their first child at 27 years or after, although the trend test was not significant (P-trend=0.15). Women who began menstruating after the age of 17 years had a 1.8-fold risk (95% CI 1.03–3.24) of gallbladder cancer, compared with those with menarche at age 13 years or younger, but the trend test was borderline significant (P-trend=0.05). Breastfeeding was associated with a reduced risk of bile duct cancer , but the association became borderline significant after adjustment for parity . Parity was not associated with biliary stones, regardless of their location in the biliary tree, age at stone diagnosis , or duration between diagnosis and interview . None of the other reproductive factors were associated with biliary stones, and there was no effect of menopausal status, age at menopause, or menstruation duration on the risk of biliary tract cancer.Of the biliary diseases examined, the strongest associations were for gallbladder cancer after adjustment for age, education, and body mass index . Parity P-interaction=0.09). Among women with stones, the effect of parity was attenuated after further adjustment for age at first birth and age at menarche . Younger age at first birth (<27 vs ⩾27) was associated with a two-fold risk, but the estimates did not increase monotonically (P-trend=0.06) among those with stones; the association was weaker among those without stones (P-trend=0.39) (P-interaction=0.07). Among women with stones, the association between age at first birth and gallbladder cancer was not considerably changed when further adjusted for parity and age at menarche. Older age at menarche was significantly associated with gallbladder cancer among women with stones , but nonsignificantly among women without stones  (P-interaction=0.86). The association between age at menarche and gallbladder cancer remained significant after further adjustment for parity and age at first birth, although the estimates did not increase monotonically.The effects of parity, age at first birth, and age at menarche on gallbladder cancer risk were more pronounced among women with stones, but the interactions were not statistically significant . Among wr=−0.66, P<0.0001), we examined their joint effect on gallbladder cancer among women with stones (Given the strong inverse correlation between parity and age at first birth among controls (h stones . The risWe observed that high parity, early age at first birth, and late age at menarche contributed to gallbladder cancer risk. It is unclear why we did not find an association between parity and biliary stones, as previous studies have shown positive associations .Given the high correlation between parity and age at first birth, it is difficult to tease apart their individual effects. Among women with stones, a strong joint effect was evident for parity and early age at first birth. This finding is biologically plausible, as younger age at first birth may reflect higher levels and longer exposure to oestrogen and progesterone (The association between late age at menarche and gallbladder cancer is consistent with the results of some previous studies , but notBreastfeeding was associated with a reduced risk of bile duct cancer, independent of stones, parity, and age at menarche, but was not associated with gallbladder cancer. Breastfeeding was related to a higher risk of gallbladder cancer in a South-American study , our finThe null effect of menopausal status is consistent with the results of previous studies  and biliIn this study, selection bias was minimal because of the population-based design, high case ascertainment, and high response rates. Misclassification of cases was minimal because of the review of diagnostic data. The assessment of stone status allowed us to evaluate cancer risk while controlling for stones. The limited exposure to OCs and hormone-replacement therapy allowed us to assess reproductive factors independently of exogenous hormones. Despite being the largest population-based study on biliary tract cancers to date, we had limited statistical power to test for interactions. In addition, our findings may have limited generalisability because of the homogeneous Shanghai population.This study revealed increased risks of gallbladder cancer associated with higher parity, younger age of first birth, and late age at menarche, primarily among those with stones, whereas there was no effect for menopausal status or OC use. The joint effects of reproductive risk factors and stones suggest the need for further study into the hormonal and inflammatory mechanisms underpinning the development of gallbladder cancer.


We examined the role of alcohol on the risk of breast cancer by the joint oestrogen receptor (ER) and progesterone receptor (PR) status of the tumour using data from two case-control studies conducted in Los Angeles County, USA. Eligible premenopausal patients were 733 women aged ≤ 40 years and first diagnosed from 1 July 1983 to 1 January 1989. Eligible postmenopausal patients were 1169 women aged 55–64 years and first diagnosed from 1 March 1987 to 31 December 89. Patients were identified by the University of Southern California Cancer Surveillance Program. Neighbourhood controls were individually matched to patients by parity  and birth date (± 3 years). ER and PR status were obtained from medical records for 424 premenopausal and 760 postmenopausal patients. The analyses included 714 premenopausal and 1091 postmenopausal control subjects. Alcohol use was generally not associated with premenopausal risk of breast cancer, regardless of hormone-receptor status. Among the postmenopausal women, those who consumed, on average, ≥ 27 g of alcohol/d experienced an odds ratio (OR) of 1.76 [95% confidence interval (CI) 1.14–2.71] for ER-positive/PR-positive breast cancer relative to women who reported no alcohol consumption. Alcohol use was less clearly associated with risk of other receptor types among postmenopausal women. These data suggest that alcohol may preferentially increase risk of ER-positive/PR-positive breast cancer in postmenopausal women. © 1999 Cancer Research Campaign