IGF1 gene and of genes encoding for the major IGF-I carriers may predict circulating levels of IGF-I and have an impact on cancer risk. We tested this hypothesis with a case–control study of 807 breast cancer patients and 1588 matched control subjects, nested within the European Prospective Investigation into Cancer and Nutrition. We genotyped 23 common single nucleotide polymorphisms in IGF1, IGFBP1, IGFBP3 and IGFALS, and measured serum levels of IGF-I and IGFBP-3 in samples of cases and controls. We found a weak but significant association of polymorphisms at the 5′ end of the IGF1 gene with breast cancer risk, particularly among women younger than 55 years, and a strong association of polymorphisms located in the 5′ end of IGFBP3 with circulating levels of IGFBP-3, which confirms previous findings. Common genetic variation in these candidate genes does not play a major role in altering breast cancer risk in Caucasians.Insulin-like growth factor I (IGF-I) stimulates cell proliferation and can enhance the development of tumours in different organs. Epidemiological studies have shown that an elevated level of circulating IGF-I is associated with increased risk of breast cancer, as well as of other cancers. Most of circulating IGF-I is bound to an acid-labile subunit and to one of six insulin-like growth factor binding proteins (IGFBPs), among which the most important are IGFBP-3 and IGFBP-1. Polymorphisms of the  Insulin-like growth factor-I (IGF-I) is a peptide growth factor that exerts mitogenic and metabolic activities that are regulators of growth, survival and cell differentiation in a number of cell and tissue types. A number of epidemiologic studies have shown a direct association between circulating IGF-I levels and breast cancer risk, especially in young women , which forms a ternary complex with an acid-labile subunit (ALS). The smaller complexes are able to pass the vascular endothelial barrier and therefore may be important modulators of IGF-I activity at the cellular level .While nutrition is an important determinant of circulating IGF-I levels , heritabFor the present study, we focused on 23 common single nucleotide polymorphisms (SNPs) that we estimated would have the highest chance of having an impact on either the expression or function of IGF-I and of molecules  involved in IGF-I transport. We conducted a case–control study of 807 breast cancer patients and 1588 matched control subjects, nested within the cohorts of the European Prospective Investigation into Cancer and Nutrition (EPIC) , and exaThe EPIC cohort consists of about 370 000 women and 150 000 men, aged 35–69 years, recruited between 1992 and 1998 in 23 research centres in 10 Western European countries. Although detailed information on ethnicity of EPIC subjects is not available, in practice, recruitment has not been conducted in large cosmopolitan urban areas; therefore, owing to the current ethnic composition of the regions involved in the study, we estimate that the vast majority (>97%) of subjects recruited in the EPIC cohort are of Caucasian origin. All EPIC study subjects provided anthropometric measurements  and extensive, standardised questionnaire information about medical history, diet, physical activity, smoking and other lifestyle factors. Women also answered questions about menstrual and reproductive history, hysterectomy, ovariectomy and use of exogenous hormones for contraception or treatment of menopausal symptoms. In addition, about 240 000 women and 140 000 men provided a blood sample.Cases of cancer occurring after recruitment into the cohort are identified through local and national cancer registries in seven of the 10 countries, and in France, Germany and Greece by a combination of contacts with national health insurances and/or active follow-up through the study subjects or their next of kin. Follow-up on vital status, to monitor the population remaining at risk for cancer, is achieved through record linkage with mortality registries. In all EPIC study centres, cancer diagnosis is confirmed through comprehensive review of pathology reports, and checks for completeness of follow-up are made periodically. A fully detailed description of the EPIC study has been published elsewhere .et al, manuscript submitted).Cases and controls from the present study were from 16 of the 23 EPIC recruitment centres, in seven of the 10 countries participating in EPIC , and were part of a larger nested case–control study on serum hormones and breast cancer risk, reported in detail elsewhere  were excluded from the study, because each of these various factors could have altered circulating hormone levels.et al).For each case subject with breast cancer, two control subjects were chosen at random from among cohort members alive and free of cancer (except non-melanoma skin cancer) at the time of diagnosis of the index case. Control subjects were matched to the cases by study centre where the subjects were enrolled in the cohort, as well as by menopausal status , age (±6 months) at enrolment, follow-up time, fasting status, time of the day of blood donation and phase of the menstrual cycle for premenopausal women , SNPper (http://snpper.chip.org/) and Frequency Finder (http://bluegenes.bsd.uchicago.edu/frequencyfinder/). We complemented database searches with literature review and, for IGFBP1 and IGFBP3, with analysis of 95 subjects from the EPIC population by denaturing high-performance liquid chromatography , and thus were likely to harbour regulatory sequences. We included only polymorphisms whose existence in Caucasians is documented, either according to literature data or to our own experimental analysis by DHPLC. All new SNPs identified in our laboratory by DHPLC searches have been deposited in dbSNP (http://www.ncbi.nlm.nih.gov/SNP). Among all polymorphisms thus identified, we retained only those with a minor allele frequency ⩾5% in Caucasians or those that result in an amino-acid change and had a minor allele frequency ⩾1%. Finally, we particularly favoured the inclusion of all polymorphisms that were previously reported in the literature to be associated with cancer risk and/or levels of circulating hormones. In total, this strategy led to a list of 26 SNPs for genotyping.To be included in the study, polymorphisms had to be located in exons (including untranslated regions), exon–intron junctions or promoter regions of a gene of interest, or otherwise should be within intronic regions that showed greater than 80% homology between human and mouse  with Puregene chemistry .μl of 2 × master mix (Applied Biosystems) in a final volume of 5 μl. The thermocycling included 50 cycles with 30 s at 95°C followed by 60 s at 60°C. PCR plates were read on an ABI PRISM 7900HT instrument (Applied Biosystems). Laboratory personnel was kept blinded to case–control status throughout the study. Genotyping call rates ranged between 95.27 and 99.44%. The distributions of genotypes of all polymorphisms were in agreement with the Hardy–Weinberg equilibrium . Repeated quality control genotypes  showed greater than 99% concordance for all assays.Genotyping was performed by the 5′ nuclease assay (TaqMan). The order of DNAs from cases and controls was randomised on PCR plates in order to assure that an equal proportion of cases and controls could be analysed simultaneously. TaqMan probes were synthesised by either Applied Biosystems, Foster City, CA, USA (with MGB chemistry) or Proligo, Paris, France (with or without LNA chemistry). Sequences of primers and probes are reported in Measurements of IGF-I and IGFBP-3 were performed in the laboratory of the Hormones and Cancer, at IARC, using enzyme-linked immunosorbent assays from Diagnostic System Laboratories . The IGF-I assays included an acid–ethanol precipitation step to eliminate IGF-I binding proteins, to avoid their interference with the IGF-I measurement. Measurements were performed on never thawed serum sample aliquots. The mean intra- and inter-batch coefficients of variation were 6.2 and 16.2% respectively for IGF-I, and 7.2 and 9.7% respectively for IGFBP-3.http://www-rcf.usc.edu/~stram/tagSNPs.html; Individual haplotype frequencies (i.e. estimated numbers of copies of haplotypes) were reconstructed using the program ‘tagSNPs’  carried by the individuals. For rare alleles, with a frequency less than 20% , only the dominant model was used. To test whether any association of gene variants with breast cancer risk could be mediated by alterations in circulating levels of IGF-I and/or IGFBP-3, these associations were also estimated with adjustment for serum peptide levels.IGF1 and IGFBP3 genes also at the level of well-delineated haplotype blocks within a gene. In all haplotype analyses, the most common haplotype was used as the reference category.Relationships of polymorphic gene variants with serum levels of IGF-I and IGFBP-3 were estimated by standard normal regression models, stratified by EPIC recruitment centre and further adjusted for age. Relationships of polymorphic variants with breast cancer risk (odds ratios) were estimated using conditional logistic regression models, applied on the matched case–control sets. Both series of analyses were performed at the level of single SNP loci, as well as at the level of haplotypes . Haplotype analyses were performed at the level of full gene loci – that is, including haplotypes based on all of the SNPs in that gene – and for the Adjustment for potentially relevant variables  did not alter significantly the results (data not shown). Exclusion of cases diagnosed within a year of blood collection also did not have an impact on the results (data not shown).http://www.broad.mit.edu/personal/jcbarret/haploview; Reconstruction of haplotype blocks within each gene was performed with Haploview  or after age at diagnosis of 55 years  were used to examine whether associations of gene variants with breast cancer risk differed between women with cancer at approximately premenopausal or postmenopausal age, and possible heterogeneity of effect between these two age groups was tested using a χIGFBP3 and IGFBP-3 level), and a prior probability of 0.1% may be appropriate when the biological knowledge and epidemiological data were both inadequate .We estimated the false positive reporting probability (FPRP) for statistically significant observations based on the methods described by vs 2.47, P=0.02) and were significantly older at first full-term pregnancy . Age at menarche did not differ between cases and controls, nor did body mass index. Serum levels of IGF-I adjusted for age and centre were not significantly different between cases and controls , nor for the subgroups subjects with cancer diagnosis before or at the age of 55 years , or after . Case subjects did show higher mean levels of serum IGFBP-3 than controls . The latter difference, however, was due mostly to the subgroup with cancer diagnosis after age 55 years , and was not clearly visible among the younger women . Details on the relationships of IGF-I and IGFBP-3 with breast cancer risk, with an extended series of 1195 beast cancer cases and 2321 control subjects, will be reported elsewhere .A total of 807 incident cases of breast cancer from the EPIC cohort and 1588 matched controls were included in our study. The mean age of study subjects at blood donation was 55 years (5th–95th percentile: 39.9–68.7 years). For cases, the mean age at diagnosis was 57 years (5th–95th percentile: 42–71 years). Based on the questionnaire data, 32% of the subjects were premenopausal at blood donation, 10% were perimenopausal or of unknown menopausal status and 58% were postmenopausal. Cases had a significantly lower number of full-term pregnancies than controls  and also a modest effect in the heterozygote (P-trend for codominant model=0.03). Interestingly, when the cases were stratified by age of onset , this reduction appears to be confined to breast cancer cases with an early age of diagnosis, at 55 years or before, and was strongest  in the homozygous G/G (P=0.005). This association was also found to be statistically significant under the dominant (P=0.01) and codominant (P=0.002) models. Linkage disequilibrium (LD) reconstruction of this region shows that there is strong LD between the rs2162679 and the two surrounding SNPs, rs35765 and rs35767. As expected, rs35767 and rs35765 also showed a reduction in breast cancer risk among the entire set, although the only statistically significant result was seen for rs35767 when testing for trend (P=0.04). However, as with rs2162679, the reduction effect was statistically significant in both rs35767 and rs35765 for the breast cancer cases diagnosed at age 55 years or before. Haplotype analysis of the IGF1 data showed only an association between one rare haplotype and cancer risk . The only other result of interest in the IGF1 gene was a relatively modest dominant effect of polymorphisms rs35765, rs35767 and rs6220, resulting in a higher mean circulating IGF-I level, and visible also in the haplotype analysis.We also noted a significant decrease in circulating IGFBP-3 levels with the rare homozygous states for both rs2162679 and rs35767 (again representing the LD between these markers). In contrast to the association of breast cancer risk, with the same markers, this effect appeared restricted to a later age of onset in our age-stratified analysis. When we performed the same analysis by using the haplotype block that spans the 5′ end of the gene and includes rs2162679 and rs35767, the same haplotype that is associated with reduced cancer risk appeared also to be strongly associated with decreased IGFBP-3 levels  (P⩽0.0001). Reflecting the strong amount of LD in this area  was insufficient to do so accurately. In view of the relative paucity of publicly available (e.g. HapMap) data, an accurate haplotype tagging approach would have required complete resequencing of the gene region in a sufficient number of subjects, in order to establish a complete catalogue of polymorphisms and to examine LD patterns between them. Such effort was beyond the scope of the present project. Nevertheless, we feel that it is unlikely that many new common polymorphisms, which were the focus of our investigation, would have been discovered by systematic resequencing.We have performed a large-scale association study, nested in the EPIC cohort, to assess the role of genetic variation of The use of a multicentric study raises the possibility of confounding by population stratification. Although 97% of EPIC subjects are estimated to be of Caucasian origin, there could still be confounding by population stratification if SNP or haplotype allele frequencies varied between the subject recruitment areas, while at the same time there was also variation in average breast cancer incidence rates. We did not observe wide differences in allele frequencies between different countries in our study (data not shown). In addition, breast cancer cases and control subjects were systematically matched for the study centre where they had been recruited into the EPIC cohorts, and we adjusted all statistical analyses of association between SNPs, hormone levels and breast cancer risk for the factor ‘recruitment centre’. We believe therefore that confounding by population stratification is not an issue in our study.IGF1. This association was particularly strong among women younger than 55 years.We have found an association with decreased breast cancer risk of a haplotype located in the 5′ part of IGFBP3. These associations are likely to be chance findings, as they are based on small numbers.No other variant among the ones we studied showed any association with cancer risk, except for two haplotypes of Our results suggest that in none of the four genes we examined were there any SNPs that had a strong impact on circulating levels of IGF-I.IGFBP1, which harbours sequences affecting gene expression, had significantly lower circulating levels of IGF-I both before and after overfeeding. Subjects carrying two copies of the A allele had lower IGF-I concentrations before overfeeding, which were further decreased after overfeeding compared to subjects carrying the rarer G allele. This may indicate that the AA genotype results in higher IGFBP-1 concentrations, which could decrease available IGF-I. We found no evidence of higher IGF-I concentrations with this polymorphism. It is possible that the previously reported association  from the transcription start site (in vitro, a construct including only the A allele of the rs2854744 SNP variant was found to increase promoter activity, suggesting that rs2854744 is the functional variant that affects IGFBP-3 transcription, and that the associations we have observed with the other polymorphisms of IGFBP3 are exclusively due to strong LD (The finding supported by the strongest statistical evidence in our study is the association of polymorphisms in the 5′ region of art site . This leThe −202 polymorphism also has been studied in relation to risk of breast cancer (IGFBP3 gene, another strong predictor of variability in IGFBP-3 levels was the same IGF1 haplotype that was associated, in the same direction, with cancer risk. This was particularly evident when the analysis was restricted to the haplotypes observed within the strong LD block in the 5′ region of the gene. It is difficult to interpret these associations, with either IGFBP-3 levels or cancer risk, as for none of these SNPs is there any evidence to suggest a possible functional role. Little appears to be known about what genomic elements are involved in the regulation of IGF1 transcription, except for a region immediately upstream of the transcription start (IGF1, between SNPs rs35767 and rs2162679, which has been frequently included in previous studies. Several epidemiological studies on the association between this polymorphism and breast cancer risk or IGF-I levels (IGF1 CA repeat has a functional role of its own, and that associations previously reported with this microsatellite are likely to reflect LD with the same SNPs that we included in our study or with other, yet unknown SNPs located in the region. In any case, the level of LD in the 5′ region of the gene is so high that an association study approach alone will not be able to single out one particular variant as that causing variation in IGFBP-3 levels and, possibly, cancer risk. Moreover, genetic variation in the promoter of IGF1 should have an impact on circulating IGFBP-3 only through a modification of circulating IGF-I, which we observe only at modest levels. Functional studies will be needed to clarify this point.Besides polymorphic variation in the IGF1, in homozygosity, and SNP rs3751893 of IGFALS – are difficult to evaluate, because of the small number of subjects carrying the alleles that showed the associations.Other polymorphic variants that were found to be associated with IGFBP-3 levels – a haplotype in the 5′ region of IGF1 gene with breast cancer risk, particularly among women younger than 55 years, and an association of polymorphisms located in the 5′ region of IGFBP3 with circulating levels of IGFBP-3. The large number of statistical tests we have performed raises the issue of potential false positives. An alternative to applying a Bonferroni's correction, which is generally too conservative because of statistical dependence between tests for multiple SNPs that are in LD, is the use of a Bayesian approach, such as the recently introduced calculation of FPRP (IGF1 SNPs we found associated with breast cancer risk, we calculated FPRPs by using a prior probability of true association. Calculated FPRPs were high even with a relatively high prior probability of 0.01 that variants would have an association with breast cancer risk or with circulating peptide levels. Our finding relating IGFBP3 SNPs with serum IGFBP-3 levels, however, is strongly backed by previous epidemiologic and functional findings and is also supported by very low P-values in our study, and resulted in very low FPRPs over a wide range of prior probabilities of true association.In summary, the main findings of our study were thus a weak but nominally significant association of a block of polymorphisms located at the 5′ end of the IGFBP3 SNPs with IGFBP-3 levels. On the other hand, we have found only weak or no associations of genetic variants in IGF1, IGFBP1, IGFBP3 and IGFALS with breast cancer risk, and further large studies will be required to confirm our findings.In conclusion, our results show a number of genetic variants associated with circulating hormone levels, including a convincing association of


CHEK2 gene is involved in cellular repair of DNA damage. The truncating mutation, *1100delCCHEK2, seems to increase the risk for breast cancer. We investigated whether the *1100delCCHEK2 mutation carrier status increases the risk for asynchronous contralateral breast cancer (CBC) and whether it interacts with radiation therapy (RT) or chemotherapy in regard to CBC risk. The germline mutation frequency was assessed in 708 women with CBC and 1395 women with unilateral breast cancer (UBC) in the Women's Environment, Cancer and Radiation Epidemiology (WECARE) Study whose first primary breast cancer was diagnosed before age 55 years and during 1985–1999. Seven women with CBC (1.0%) and 10 women with UBC (0.7%) were *1100delCCHEK2 variant carriers (rate ratio (RR)=1.8, 95% confidence interval (CI)=0.6–5.4 for CBC vs UBC). Carriers who received RT for their first breast cancer, compared with non-carriers not treated with RT, had an RR of developing CBC of 2.6 (95% CI=0.8–8.7). We found no significant associations between the *1100delCCHEK2 mutation and CBC overall or among those treated with RT. However, the sampling variability was such that modest increases in risk could not be excluded. Nonetheless, because this is a rare mutation, it is unlikely to explain a major fraction of CBC in the population.The protein encoded by the  CHEK2 gene codes for a protein that is involved in cell-cycle control and DNA repair. A protein-truncating mutation in the gene, 1100delC, was associated with a two-fold increase in breast cancer risk in the CHEK2 Breast Cancer Case–Control Consortium pooling project, which comprised 10 860 breast cancer cases and 9065 controls from 10 studies in five countries (The *1100delCCHEK2 variant among women with and without breast cancer varies among countries (*1100delCCHEK2 mutation are hypothesised to have an increased sensitivity to exposures such as radiation therapy (RT) and chemotherapy that cause DNA double-strand breaks, as the CHEK2 regulates cell-cycle checkpoint response pathways that respond to such damage  Study (CHEK2 and having received RT or chemotherapy for breast cancer.The ries The . The preh damage . We invein situ) at least 1 year later, while residing in the same reporting area at the time of diagnosis of both cancers with no history of a prior cancer diagnosis or an intervening cancer diagnosis between their first and second primary breast cancers. All first primary breast cancers were invasive without distant metastases and were diagnosed between 1 January 1985 and 31 December 1999. The second primary breast cancer had to have been diagnosed no later than 31 December 2001. All women were alive at the time of contact. Control women had to fulfil the same eligibility criteria as cases except that they had no CBC during the interval between the matched case woman's first and second breast cancers (‘at-risk’ period) and had no prophylactic mastectomy during the ‘at-risk’ period. The reference date for case women was the date of diagnosis of their second primary breast cancer, whereas the corresponding reference date for control women was defined as their date of diagnosis of primary breast cancer plus the ‘at-risk’ period for their matched case. Two UBC controls were individually matched to each CBC case on year of birth (5-year strata), year of diagnosis (4-year strata), registry region and race, and counter-matched on RT (as designated by the cancer registry), that is, each triplet comprised two women who had received RT and one woman who had not. Counter-matching was used to improve the statistical efficiency of the study design  obtained from the WECARE Study . By the use of methods previously described (*1100delCCHEK2 (four controls excluded).Among 998 CBC patients (cases) and 2112 UBC controls who were eligible and approached to participate in the study, 708 (71%) cases and 1399 (66%) controls were enrolled. Matching was successfully carried out for 694 triplets (matching criteria were relaxed on region or race/ethnicity for 17 triplets) and 11 case–control pairs, whereas three cases had no matched controls. All 708 cases and 1395 of the controls (99.7%) were genotyped for *1100delCCHEK2, we developed an assay based on ligation-dependent amplification  and 2 fmol of each oligonucleotide were added, and the samples were then heated at 95°C for 1 min and incubated at 37°C for 16 h in a total volume of 8 μl. Ligation of the hybridised oligonucleotides was achieved by the addition of 32 μl of a ligase mix containing 3 μl ligase buffer A (MRC-Holland), 3 μl ligase buffer B (MRC-Holland), 1 U Taq DNA ligase  and incubation at 54°C for 15 min. Ligation products were amplified with primers 5′-TATGTAAAACGACGGCCAGT-3′ and 5′-TATTATAGGGCGAATTGGGT-3′. Hybridisation, ligation and PCR were performed in 96-well plates, which contained two positive controls (DNA with *1100delCCHEK2), two negative controls (DNA without *1100delCCHEK2) and two ligation controls (without DNA). The PCR products were resolved by agarose gel electrophoresis.For detection of fication . The seqThe reliability of the ligation assay was confirmed by re-analysis of 248 DNA samples from WECARE Study participants using denaturing gradient gel electrophoresis (DGGE), as described previously (BRCA1/2 carrier status (deleterious BRCA1/2 mutations yes/no/unknown). The TPHREG procedure in SAS release 9.1  was used for the statistical analyses. Further information about the SAS programming used for counter-matched WECARE data is provided in the Rate ratios (RRs) and corresponding 95% confidence intervals (CIs) were estimated by univariate conditional logistic regression models accounting for the counter-matched design  carriers were found among the 1395 women with UBC. Characteristics of carriers and non-carriers among cases and controls are shown in The mean length of time since first breast cancer diagnosis was 5.0 years (range: 1–15.6 years) for the total study population. Seven (1.0%) carriers of the *1100delCCHEK2 mutation and risk of developing CBC was not statistically significant   as were the RRs for the mutation in combination with RT or chemotherapy (The association between being carrier of the 0.6–5.1) . Compareotherapy .Results of analyses restricted to non-Hispanic whites (649 women with CBC and 1286 women with UBC) were similar to those for all study subjects combined.*1100delCCHEK2 mutation carriers among 708 women with CBC (1.0%) and 1395 women with UBC (0.7%). Several studies have shown higher frequencies (ranging from 1.7 to 14.3%) of mutation among women with bilateral breast cancer (BRCA1/2 mutation cases (In this population-based multicentre study, we found a low prevalence of *1100delCCHEK2 allele is hypothesised to interact with environmental factors to promote breast tumorigenesis. Ionising radiation is of particular interest because it induces DNA damage, and the protein of the CHEK2 gene is a kinase that plays a role in response to such damage by delaying cell-cycle progression to facilitate DNA repair or even induce cell death (*1100delCCHEK2 mutation are, in general, not characterised by a distinct enhanced chromosomal radiosensitivity, this conclusion was based on only four breast cancer cases with the mutation who were compared with healthy controls without the mutation (The ll death . The gerBRCA1 and BRCA2 genes, we were able to adjust the analyses for BRCA carrier status. Patients included in the study survived varying lengths of time from reference date to interview; however, cases and controls within each triplet survived approximately the same length of time. The prevalence of the mutation may have been affected if carriers of the *1100delCCHEK2 mutation had a worse survival than non-carriers. Unfortunately, we could not address this question, as we had no information on carrier status among the non-survivors who were not included in the study. Previous studies have shown a worse recurrence-free survival (The WECARE Study was designed to investigate interactions between genes and exposures that cause DNA double-strand breaks in a study population enriched with hereditary breast cancer cases, owing to the restriction to early-onset breast cancer, and a population heavily exposed to ionising radiation and chemotherapeutic drugs. Detailed information on RT and chemotherapy was obtained by a thorough review of medical records. As DNA samples from WECARE Study participants have also been screened for mutations in *1100delCCHEK2 mutation in this large population-based sample of early-onset cases of CBC and UBC. The small number of carriers reduced our ability to study the effects of the mutation on the development of CBC; and accordingly, we could not exclude the possibility that the *1100delCCHEK2 mutation is associated with modest increases in risk. In contrast, the rarity of the *1100delCCHEK2 mutation indicates that it plays a rather limited role in the aetiology of CBC in the population.Notably, we found a low carrier frequency for the


P for interaction=0.0036). The incidence rates per 100 000 person-years for those who consumed >3 drinks per day were similar in men (77.6) and women (75.3). The higher HRs observed in women resulted from lower incidence rates in the referent group: women (14.7), men (34.4). In summary, drinking >3 alcoholic beverages per day was associated with increased risk in men and women, but consumption of up to one drink per day may be associated with reduced risk relative to non-drinking.We investigated the relation between head and neck cancer risk and alcohol consumption in the NIH-AARP Diet and Health Study. During 2 203 500 person-years of follow-up, 611 men and 183 women developed head and neck cancer. With moderate drinking  as the referent group, non-drinkers showed an increased risk of head and neck cancer (men: hazard ratio (HR) 1.68, 95% confidence interval (95% CI) 1.37–2.06; women: 1.46, 1.02–2.08). Among male and female alcohol drinkers, we observed a significant dose–response relationship between alcohol consumption and risk. The HR for consuming >3 drinks per day was significantly higher in women  than in men (1.48, 1.15–1.90;  Each year, more than 600 000 individuals are diagnosed with squamous cancers of the head and neck worldwide (Alcohol use has consistently been indicated as a risk factor for head and neck cancer . Few proAlthough men are 2–7 times more likely to develop these cancers than women (depending on sub-site)  and subjects with cancer at baseline (n=51 219), calorie intake more than two interquartile ranges from the mean (n=4419), who died or were diagnosed with cancer on the first day of follow-up (n=10), and who failed to provide information on alcohol use (n=2039). The resulting analytical cohort comprised 492 960 participants.The establishment and recruitment procedures of the NIH-AARP Diet and Health Study have been described  and usual smoking dose to construct a categorical variable (smoke-quit-dose): never smokers, former smokers who smoked less than one pack per day, former smokers who smoked more than one pack per day, current smokers who smoked less than one pack per day, and current smokers who smoked more than one pack per day. We lacked information on smoking duration.χ2 test. Age-adjusted incidence rates were calculated as in Analyses were performed with SAS version 8.2. Potential confounders were tabulated by alcohol intake. Hazard ratios (HR) and 95% confidence intervals (95% CI) were calculated using Cox proportional hazards regression . We testAs shown in P=0.0036, 3 df) and present gender stratified estimates throughout. The age-standardised incidence rate per 100 000 person-years among moderate drinking women (up to one drink per day)  was lower than moderate drinking  men. Among drinkers, increasing alcohol consumption was significantly associated with increased risk. The incidence rates in those who drank >3 alcoholic drinks per day were similar in men and women   . In age-.1–92.1) , men and.1–92.1) .When examined by alcohol type, consuming more than three drinks of liquor or beer per day was associated with increased risk relative to moderate consumption of these beverages in both men and women .We also investigated the association between alcohol intake and head and neck cancer sub-sites. Women who drank more than three alcoholic drinks per day were at significantly increased risk of oral cavity cancer  and nonsignificantly increased risk of oropharyngeal and hypopharyngeal cancer  and laryngeal cancer  relative to moderate alcohol drinkers. Men who drank more than three alcoholic drinks per day were at significantly increased risk for cancers of the oral cavity , oropharynx and hypopharynx , but not the larynx compared to moderate drinkers  .In both age-adjusted models , non-driWe performed several sensitivity analyses. Years of education and alcohol intake were positively correlated (Pearson correlation of 0.12); so we stratified our analysis on years of education. The stratified estimates for non-drinkers relative to moderate drinkers were 1.70 (1.21–2.37) for men and 2.57 (1.20–5.53) for women who had a college education or greater and 1.76 (1.35–2.27) for men and 1.35 (0.89–2.03) for women without a college education. We also stratified by smoking status . Non-drinking men and women were at elevated risk relative to moderate drinkers among current , former , and never smokers . We also excluded those who reported poor health on the questionnaire. After excluding 34 580 men and 25 013 women who reported poor health at baseline, non-drinking men  and non-drinking women  were at increased risk relative to moderate drinkers. Similarly, after excluding the first 2 years of follow-up, non-drinking participants remained at increased risk .In a large prospective study, we investigated the association between alcohol and head and neck cancer in men and women separately. Consistent with previous studies , predomiIn this study, although the absolute age-standardised incidence rates among those who drank 1–3 alcoholic drinks per day and those who drank >3 alcoholic drinks per day were similar in men and women, the HR for the association between alcohol intake and head and neck cancer risk were significantly higher for women than for men. Similar increased risks among women were observed in most previous studies of this association . Years of education, as with the other covariates used for adjustment, did not attenuate the protective association. We additionally explored SES as a confounder using median income of the census tract in which the participant lived. Inclusion of this variable in the multivariate analysis did not alter risk estimates (data shown). We also examined possible effect modification by smoking status, as health conscious individuals may be more likely to drink moderate quantities of alcohol and less likely to smoke. Although the risk estimates for moderate alcohol consumption were similar among current, former, and never smokers, few cases with head and neck cancer were non-smokers (95 men and 20 women), limiting the power to investigate this stratum.This study has a number of strengths. It is a large prospective analysis with exposure information collected before cancer diagnosis. We present models separately for men and women and we examined the associations for both total head and neck cancer and by sub-sites. To limit confounding, we adjusted our models for most of the major risk factors for head and neck cancer, including cigarette smoking, a strong risk factor that is associated with alcohol use. Furthermore, the large number of participants who consumed less than one alcoholic drink per day allowed us to investigate the association between moderate drinking and risk. This study also had several limitations. We lacked information on participant income and occupation, factors that could confound the association between alcohol intake and risk. We also lacked information about past alcohol consumption and smoking duration.In conclusion, among alcohol drinkers, increasing consumption of alcohol was associated with increased risk of head and neck cancer in both men and women. Although consumption of >3 drinks per day had a bigger HR for head and neck cancer in women than in men, incidence rates for men and women in this category were similar. Men and women who reported consuming less than one drink per day had lower rates of head and neck cancer than men and women who reported no alcohol consumption. The observed association between increasing alcohol consumption and increased head and neck cancer in men and women is consistent with many previous studies. In contrast, the observed protective association with drinking up to one alcoholic beverage per day should be interpreted cautiously and must be investigated further, particularly in prospective studies that collect information on lifetime use of alcohol.


Bisphosphanates are used primarily for the prevention and treatment of osteoporosis, and are also indicated for osseous complications of malignancy. In addition to their bone resorption properties, the most commonly used nitrogen-containing bisphosphonate compounds also inhibit protein prenylation, and thus may exert anti-tumour properties.To evaluate whether the use of these drugs may be associated with cancer, specifically breast cancer, we conducted a population-based case–control study in Wisconsin from 2003 to 2006. Participants included 2936 incident invasive breast cancer cases and 2975 population controls aged <70 years. Bisphosphonate use and potential confounders were assessed by interview.P-trend=0.01). Risk reduction was observed in women who were not obese (P-interaction=0.005).Using multivariable logistic regression, the odds ratio for breast cancer in current bisphosphonate users compared with non-users was 0.67 . Increasing duration of use was associated with a greater reduction in risk (These results are suggestive of an additional benefit of the common use of bisphosphonates, in this instance, the reduction in breast cancer risk. Bisphosphonates are commonly prescribed for the prevention and treatment of osteoporosis . By inhiWomen aged 20–69 years with a new diagnosis of invasive breast cancer from 2003 to 2006 were identified from Wisconsin's mandatory cancer registry , age at first birth , first-degree family history of breast cancer, body mass index 1 year before reference date , menopausal status, age at menopause , screening mammograms over the past 5 years , a physician diagnosis of osteoporosis, smoking status, and height change from the age of 18 years. Effect modification was evaluated by inclusion of cross-product interaction terms in logistic models. Owing to missing or incomplete information on osteoporosis medications or their indications, 26 breast cancer cases and 29 control women were excluded, leaving 2936 cases and 2975 controls for analysis.Multivariable logistic regression models were used to estimate odds ratios and 95% confidence intervals and tests for linear trend across ordinal values of categorical variables. Current use of bisphosphonates was defined as use in the year before the reference year, and former users had taken bisphosphonates at any time before this period. Duration of use was the sum of all periods of bisphosphonate use in the year before the reference period. Analyses were adjusted for potentially confounding variables selected The average age at the reference date for breast cancer cases was 54.2 years (s.d.=8.9 years) and for controls it was 54.8 years (8.9 years). As expected, women with breast cancer were more likely than controls to have a family history of breast cancer, a college degree, to be older at first birth or nulliparous, to have gained more weight since the age of 18 years, and to use postmenopausal hormones; controls were more likely to report a physician's diagnosis of osteoporosis (data not shown).P-trend=0.01).Bisphosphonates were more commonly used by controls than by cases; the multivariate-adjusted odds ratio for ever use was 0.70 . The inverse effect of bisphosphonates seemed limited to women who were not obese , height loss, and physician-diagnosed osteoporosis. There was a suggestion that use of bisphosphonates was associated with a reduced risk of breast cancer only among women reporting symptoms of bone loss , but the differences were not statistically significant ; therefore, we could not evaluate specific types of preparations that have very different mechanisms of action (In summary, this study provides new evidence that the use of bisphosphonates is associated with a potentially important reduction in breast cancer risk. Because we were able to account for important confounders, these findings may reflect real benefits due to the anti-tumour mechanisms of these medications.


Waist circumference was a significant anthropometric factor for breast cancer in both pre- and postmenopausal women. Furthermore, the associations of leptin with breast cancer risk remained after adjustment for obesity indices. These results suggest that leptin may have an independent role in breast tumorigenesis. Regardless of the impact of circulating leptin, more research is needed to elucidate molecular mechanisms and local leptin levels that are critical for the development of breast cancers.The present case–control study was to investigate the relationships of plasma leptin level and anthropometric measures of adiposity with the risk of breast cancer. Questionnaire information, anthropometric measures and blood samples were taken before treatment from 297 incident cases with breast cancer and 593 controls admitted for health examination at the Tri-Service General Hospital, Taipei, between 2004 and 2006. Plasma levels of leptin were measured by RIA. Logistic regression analysis was used to estimate odds ratios (ORs) and 95% confidence intervals (CIs) for assessing the associations. Overall, higher leptin concentrations were significantly associated with an increased risk of breast cancer (OR (95% CI) for top  One control subject with inadequate specimen was excluded, resulting in 297 cases and 593 controls included in this study.This case–control study was conducted at the Tri-Service General Hospital, Taipei, Taiwan, from January 2004 to November 2006. To account for the notions that type 2 diabetes and metabolic syndrome (MS) have been associated with increased serum leptin levels and elevated risk of breast cancer , as an indicator of generalised obesity, was calculated as weight in kilograms divided by the square of height in metres. In addition, the measurements of waist and hip circumferences were performed by trained clinical staff using standardised techniques. Waist girth was determined with a measuring tape placed horizontally around the midpoint between the iliac crest and lower margin of the ribs. Hip girth was the maximum circumference around the buttocks posteriorly and the symphysis pubis anteriorly. As a result, waist circumference (WC) and waist-to-hip ratio (WHR) represent a measure of central adiposity. Plasma leptin concentrations were measured in a single run using a commercially available RIA kit  according to the manufacturer's instructions. The sensitivity of this assay is 0.5 ng mlt-test. Spearman correlation coefficients were used to examine cross-sectional relationships between leptin and anthropometric measures of adiposity. In addition, conditional logistic regression, which preserves the matching of cases and controls, was used to estimate odds ratios (ORs) and 95% confidence intervals (CIs) for the associations between plasma leptin levels and anthropometric measures and breast cancer risk.Differences between cases and controls in age at menarche, age at first full-term pregnancy (FFTP), age at menopause and parity numbers were tested using the Student's vs 13.7 years), age at FFTP (26.7 vs 26.5 years), age at menopause (48.9 vs 48.5 years) and parity number (2.0 vs 2.0). By contrast, there were statistically significant differences between cases and controls for anthropometric measures of adiposity. The average BMI, WC and WHR were higher in breast cancer cases than in control subjects .The baseline characteristics of cases and controls are summarised in r=0.59, P<0.0001), WC  and WHR . The mean (±s.d.) levels of leptin in cases with invasive breast cancer, patients with carcinoma in situ and control subjects were 10.4(±7.0), 8.7(±5.3) and 8.4 (±5.3 ng ml−1), respectively. We further categorised plasma leptin either as high or as low levels (based on the 75 percentile in the control group) or into three groups  when appropriate for subsequent analyses. Ptrend=0.0091). In subgroup analyses according to menopausal status, a similar pattern regarding breast cancer risk associated with higher levels of leptin was observed in postmenopausal women. Additional adjustment for BMI and WC slightly attenuated these associations. Restricting analysis to cases with invasive cancers and control subjects did not reveal any other remarkable discrepancies in results (data not shown). Concerning the roles of general and central obesity, we found that increasing BMI values were associated with reduced risk of breast cancer in premenopausal women after adjustment for circulating leptin levels , whereas an increasing WC was associated with increased risk of breast cancer in both pre- and postmenopausal women when an adjustment for leptin concentrations was made (ORs (95% CI) for the highest vs the lowest tertile in pre- and postmenopausal women were 3.54 (1.55–8.10) and 2.85 (1.19–6.84), respectively). Accordingly, WC was a significant anthropometric factor for breast cancer. Thus, we further evaluate the possible joint effect of WC and leptin on breast cancer risk for the whole group and separately for pre- and postmenopausal women  and in that of epidemiological studies (range: 8–41 ng ml−1). Furthermore, regardless of the impact of circulating leptin, recent studies suggest that carcinogenesis could be induced by an overabundance of locally produced leptin. The analysis of approximately 300 biopsies revealed that leptin is overexpressed in breast cancer, whereas it is absent or expressed at very low levels in normal epithelium or benign tumours  (vs 26.5 days), and this average duration was not significantly associated with circulating leptin levels . In addition, our data showed that there was no significant difference in leptin concentrations between women who received postmenopausal HRT (51 cases and 145 controls) and those who were non-HRT users (74 cases and 114 controls) . Thus, the phase of menstrual cycle in premenopausal women and HRT in postmenopausal women may not affect the case–control comparisons for leptin concentrations in this study.The limitations of our study design warrant discussion. There was only one-time measurement of leptin concentration in this study. However, a single measurement of circulating leptin levels appears to reflect an individual's long-term levels of leptin quite well, because even in samples taken 4 years apart, intraclass correlation was high (in vitro studies provide the background biological context for considering a role for leptin in the pathogenesis of breast cancer. Further investigation is warranted to confirm and expand on this empirical finding.In conclusion, the converging lines of evidence from this study and earlier


The possible influence of induced abortion on breast cancer risk was assessed in a cohort of 267 040 women enrolled in a randomised trial of breast self-examination in Shanghai, China. Based on answers to a baseline questionnaire, subsequent breast cancer risk was not significantly associated with ever having an induced abortion. After adjustment for potential confounders, the relative risk estimate was 1.06 , and there was no trend in risk with number of abortions. Analysis of data from more detailed interviews of 652 cases and 694 controls from the cohort yielded similar results. There was also no overall increase in risk in women with induced abortion after first birth. Few women had undergone an abortion after 13 weeks gestation or before their first child. Although increases in risk were observed in such women, they were not statistically significant and could have been due to recall bias. Abortions as they have been performed in China are not an important cause of breast cancer.British Journal of Cancer (2002) 87, 977–981. doi:10.1038/sj.bjc.6600603www.bjcancer.comCancer Research UK© 2002  During the early months of a pregnancy, the mammary epithelium undergoes massive proliferation and then this is followed by differentiation in preparation for lactation . It has Induced abortions have been widely used in China since the 1970s as part of the national family planning programme, aimed at achieving the goal of one child per family. Abortions are performed primarily on married women to limit family size. These are socially acceptable and readily reported  in female textile workers in Shanghai . All curThere were 292 503 women listed on the original roster, and after eliminating women who were ineligible, had left the STIB by the time of the baseline activities, did not complete the baseline questionnaire, or who had breast cancer before the beginning of the BSE study, 267 040 women remained in the BSE trial, and constituted the population for this cohort study.As part of the BSE trial, all new breast cancer cases were identified by factory medical workers at each factory. This was supplemented by checking the records of the STIB tumour and death registry, which passively receives reports of all incident cancer cases from each factory annually, and by periodic review of the records of the Shanghai Cancer Registry. When a case was identified, her name, age, diagnosis date, hospital, address and the BSE study identification number were obtained. From September 1, 1989 to August 31, 1995, 702 cases of histologically confirmed breast cancer were identified and included as incident cases for this study.The same 702 incident cases of breast cancer were also eligible for this nested case–control study. Controls were randomly selected from the BSE cohort. They were frequency matched to the cases using 34 exact year of birth (1925–1958) categories. Twice the number of controls as the number of cases born in each calendar birth year was selected. The women with randomly assigned odd numbers were selected as primary controls and the women with even numbers were asked to participate if the attempt to access the first control failed. In all but 42 instances first controls were interviewed and included in the study.Data collection was carried out between September 1, 1995 and February 28, 1996. Fourteen interviewers were recruited from retired female nurses at three hospitals affiliated with the STIB. Cases and controls were interviewed in person in the women's homes. The interview took approximately 1 h and requested detailed information on the major known and suspected risk factors for breast cancer. Detailed information on each abortion was collected, including the woman's age at time of the procedure, whether it was before or after other pregnancies, and weeks of gestation.Informed consent was obtained from each woman prior to interview. The study was approved by the Institutional Review Boards of the Fred Hutchinson Cancer Research Center and the Station for Prevention and Treatment of Cancer of the Shanghai Textile Industry Bureau, in accordance with an assurance filed with the Office for the Protection from Research Risks (OPRR) of the National Institutes of Health.In exploratory analyses unconditional multiple logistic regression models were used to calculate odds ratios  as estimates of relative risk. To evaluate possible confounding factors, age plus individual risk factors were entered sequentially into logistic regression models as categorical variables, and those that altered the age adjusted relative risk estimate in relation to induced abortion by over 5% were included in the final model. The significance of trends in risk with level of exposure was calculated by assigning scores to the categorised levels of exposure and treating the scores as a continuous variable in the regression models. To evaluate possible interactions between induced abortion and other risk factors, the risk of breast cancer in relation to induced abortion was computed by stratified analysis. Likelihood ratio tests were performed to test interactions between induced abortion and all other variables included in the final analyses. The risk of breast cancer in relation to spontaneous abortion was also estimated after adjustment for potential confounding variables, using the same procedures as for induced abortion. Logistic regression models were used in order to facilitate direct comparisons of the odds ratios derived from the case–control study.Odds ratios and 95% confidence intervals obtained from unconditional logistic regression models were used to estimate relative risks. Age was included in all logistic models. The same procedures that were used in the cohort study were employed to identify and control for potential confounding factors, and to test for trends and interactions. The risk of breast cancer in relation to spontaneous abortion was similarly estimated after adjustment for potential sources of confounding.Some items in the case–control study questionnaire were comparable to those in the baseline questionnaire, except that they were obtained nearly 6 years later. To evaluate possible recall bias in the case–control study, the information collected for that study on history of induced abortion and other variables of interest was compared to the corresponding data obtained on the baseline questionnaire after accounting for possible differences due to the different times of data collection.Compared to women having a first birth before age 20, nulliparous women had a significantly increased risk of breast cancer  and there was a significant trend in risk with age at first birth. Risk was inversely related to number of live births and reduced in women who had ever breast fed . Women who were post-menopausal at baseline were at slightly lower risk than pre-menopausal women of the same age. Risk was increased in women having a history of a breast lump  and a family history of breast cancer . No associations of risk were observed with: marital status, age at menarche, history of a stillbirth, tubal ligation, hysterectomy or oophorectomy, or use of oral or injectable contraceptives, an IUD, cigarettes, or alcohol.After adjustment for age and age at first live birth, no other variable altered risk in relation to an induced abortion by more than 5% and only these variables were included in the final model. After adjustment for age, no other variable altered the relative risk in relation to a spontaneous abortion by more than 5%, but age at first birth was included in the final model to make it comparable to the model for the risk in relation to induced abortion.As shown in Table 1The relative risk estimates in women under the age of 50 years, and aged 50 years or over, were 1.02  and 1.07  for women with an induced abortion, and 1.31  and 1.00  for women with a spontaneous abortion, respectively. No significant interactions were observed between history of an induced abortion and any of the other variables considered.Of the 702 eligible cases identified in the cohort, 652 (93%) were interviewed, 36 cases were deceased, eight were not located, and six refused to be interviewed. Of the 744 selected controls, 694 (93%) were successfully interviewed, four were deceased, 38 were not located and eight refused to be interviewed.Most of the associations with risk that were observed in the cohort study were also observed in the case–control study. These include relationships to nulliparity, age of birth of first child, number of live births, having ever breast fed, menopausal status, history of benign breast disease, and family history of breast cancer. In addition, risk was associated with several variables not available from the baseline questionnaire, including more than 12 years of schooling, body mass index at age 20 years and, unexpectedly, a first pregnancy ending in a stillbirth . As with results based on the baseline questionnaire, risk as estimated from the case–control study was not significantly altered in relation to a history of ever having a stillbirth, tubal ligation, hysterectomy or oophorectomy, or use of oral or injectable contraceptive or an IUD, alcohol drinking, or history of smoking. It was also not associated with level of physical activity at work, or the smoking history of the woman's husband.As shown in Table 2There are no significant trends in risk with number of induced abortions, age at first or last induced abortion, or years since first or last abortion. Unlike results from the cohort study, there is a possible increased risk in relation to an induced abortion before a woman's first birth, but the relative risk estimate is based on small numbers of women and has a wide confidence interval. There is a non-significant increase in risk in relation to duration of gestation before either first or last induced abortion.When the analysis was restricted to women aged under 50 years, no significant increases in risk in relation to induced abortion were evident, but a possible increased risk was seen in women having had an induced abortion before their first birth  and in woman whose first  or last  induced abortion occurred after 13 weeks gestation. There were insufficient numbers of women age 50 years or over with induced abortions before their first birth for meaningful analysis. These women did have a non-significant increase in risk in relation to 13 or more weeks at gestation before first  and last  induced abortion.Only seven cases and one control had their first abortion after 9 weeks and before their first birth, giving a risk of 7.38  relative to women with no induced abortion (320 cases and 338 controls). The relative risk for first abortion after 13 weeks and before first birth could not be estimated because of small numbers of exposed women. There was no significant alteration in risk in women having an induced abortion after their first birth regardless of length of gestation.The relative risk of breast cancer in relation to an induced abortion in women was estimated separately in women with and without various risk factors for breast cancer, and no significant interactions between induced abortion and other variables were observed.Women who had experienced a spontaneous abortion were not at increased risk , either before or after the age of 50 years. Risk was also not increased in relation to a spontaneously aborted first pregnancy .A comparison of data on induced abortion from the case–control and baseline questionnaires is shown in Table 3To assess the possible effect of recall bias on the results, the relative risk was estimated using various combinations of the data from the cohort and the case–control studies to define women who had ever undergone an induced abortion. These analyses included 652 cases and 694 controls shown in The present cohort study did not show an excess risk of breast cancer in relation to ever having an induced abortion, or in relation to number of such abortions, in women either under or over the age of 50 years.Since the present study was conducted in a population with low risk of breast cancer, if an effect of an abortion on risk is additive the present study would have been more likely to have detected the effect than studies in high-risk populations.The increase in risk in relation to induced abortion observed in most other studies was small , and resIncomplete information on induced abortion irrespective of the development of breast cancer is another possible explanation for the absence of a relationship between breast cancer and induced abortion in either the present cohort or case–control study. However, unlike previous studies, such as those carried out in the United States, where induced abortion was legalized only after 1970 and tends not to be socially acceptable, our investigations were conducted where induced abortion has been legal for over 50 years, encouraged by public policy and widely acceptable. Under such circumstances, under-reporting of most induced abortions would not likely be an important problem. In a population-based case–control study in Shanghai (in which no association between breast cancer and induced abortions was observed), 66% of the study participants resported having had an induced abortion  in women who had ever undergone an induced abortion (Differential reporting of induced abortions by cases and controls is a problem that has been identified in case–control studies of induced abortion and breast cancer in other countries (If induced abortion was related to poorly differentiated breast cancer, with a poor prognosis, deceased cases would be more likely to have a history of induced abortion than those who were interviewed, and the risk in relation to induced abortion would have been underestimated (A full-term pregnancy first causes mammary cell proliferation and then differentiation, thus presumably reducing susceptibility to carcinogenesis. It is speculated that an early interruption of a pregnancy may lead to enhanced proliferation of breast tissue without subsequent differentiation, and hence to increased susceptibility to carcinogenic change. If this is true, then risk of breast cancer would be expected to increase with the gestational length prior to an abortion, and this effect might be particularly evident in women without a prior full term pregnancy. Compatible with this hypothesis and with a prior study in Denmark (Although our study could not exclude the possibility of an increased risk in women having an induced abortion before their first birth and after a relatively long period of gestation, abortions in China are generally performed early and primarily to limit family size, and are not an important cause of breast cancer. They are not a likely explanation for the recent increase in rates of breast cancer in that country (


Diagnoses of SARDs, including rheumatoid arthritis , systemic lupus erythematosus , Sjogren's syndrome (n=374), systemic sclerosis (n=128), and dermatomyositis (n=31), were determined from claim files for individuals from age 65 years to 1 year before selection. Associations of SARD diagnoses with breast cancer, overall and by oestrogen receptor (ER) expression, were assessed using odds ratio (OR) estimates from multivariable logistic regression models. The women diagnosed with RA were less likely to develop breast cancer =0.82–0.93). The risk reduction did not differ by tumour ER-status . The breast cancer risk was not associated with any of the other SARDs, except for a risk reduction of ER-negative cases  among women with SLE. These findings suggest that systemic inflammation may affect breast epithelial neoplasia.Systemic autoimmune rheumatic diseases (SARDs) are chronic inflammatory and immuno-modulatory conditions that have been suggested to affect cancer risk. Using the Surveillance, Epidemiology and End Results–Medicare-linked database, women aged 67–99 years and diagnosed with incident breast cancer in 1993–2002 ( Breast cancer (BC), the commonest malignancy and the second leading cause of cancer death among American women , is heteSeveral studies of SARDs and breast cancer risk have had inconsistent results –Medicare-linked database to evaluate whether a history of SARD diagnoses was associated with BC risk overall and by ER status of the tumour.The SEER–Medicare data linkage, a collaboration of the National Cancer Institute and the Center for Medicare and Medicaid Services, was created by linking two large population-based data sources: the SEER cancer registry system and the Medicare enrolment and claim files. Detailed description of the database is available elsewhere .From the SEER database, we included all women with incident primary invasive adenocarcinoma of the breast  who were diagnosed in 1993–2002, with no previous cancer of any type; BC cases diagnosed only at autopsy or by death certificate were excluded.Female controls who were alive and cancer free as of 1 July of the calendar year of case selection were selected at random, with replacement, from the 5% sample of the Medicare beneficiaries who resided in SEER areas. To ensure availability of claims data, cases and controls had to be 67–99 years of age, selected in 1993 or later, and have at least 12 months of simultaneous part A and B coverage (and no HMO coverage) before the selection date. Controls were frequency matched in 1 : 1 ratio to cases according to the calendar year of diagnosis and age in three categories . Women who became BC cases could be selected as controls until they were diagnosed with the cancer.Participants were considered to have SARDs if they had at least one inpatient or two outpatient/physician claims  for any of the following diagnoses: RA , SLE (710.0), systemic sclerosis (710.1), Sjogren's syndrome (710.2), or dermatomyositis (710.3). Women who met the definition for more than one condition were included in a separate category, multiple SARDs. History of SARDs was ascertained from Medicare claim files, up to 12 months before case–control selection date.Variables assessed as potential confounders of associations between BC and SARDs included age, race, socio-economic status, region of residence, history of mammography, number of physician visits 12–24 months before selection, and earlier use of immunosuppressive medications. We used the 1990 census median annual household income in the study participants’ zip code of residence as a proxy measure of individuals’ socioeconomic status. The regions of residence, based on the location of the SEER registry, were categorised as western, northeastern, north-central, and southern . HistoryWe used unconditional logistic regression to calculate odds ratios (ORs) and 95% confidence intervals (CIs) for the association between BC risk and SARDs . We computed the variance of the OR estimates using a robust variance estimator  to adjusIn evaluating BC risk by ER status, we excluded cases with unknown ER expression. We used polytomous logistic regression to estimate ORs and 95% CIs with a robust variance estimator comparing ER-positive cases and ER-negative cases with the cancer-free controls (vs <75) and race (whites vs others), to test whether these associations are modified by age or race.All final models were adjusted for age and year of selection (matching variables), race, region of residence, median-household income by zip code of residence, history of mammography, and history of immunosuppressive therapy as defined earlier. We also stratified all the models by year of selection in three groups  to investigate whether the estimated associations were affected by secular trends in coding accuracy, mammography screening, or standard of care. In the final models that evaluated overall BC risk associated with SARDs , we added two interaction terms, age , and number of visits to physicians from 12 to 24 months before case–control selection . However, more cases than controls were white (88.4 vs 83.7%), and had history of mammography 12–48 months before selection (50.5 vs 45.4%). Patterns were similar for ER-positive and ER-negative cases, with the exception of race, for which more ER-negative cases than controls were black (9.4 vs 7.7%)  had known ER status, of whom 84.9% were ER-positive. As shown in vs 7.7%) .vs 21.4%), have tumours of unknown grade (25.3 vs 12.3%), have non-specific morphology (10.8 vs 1.5%), and no recorded stage (5.6 vs 0.7%). Although statistically significant, absolute differences between cases with known and unknown ER status with respect to age, race, and income were small (<5%)  .n=3033 in BC cases and 3396 in the controls) met the study definition of at least one SARD condition. Among them, 95.2% had only one condition. Of those who had been diagnosed with more than one condition, 84.6% had RA.Of all participants, 4.0% . When comparing individual SARD cases with no SARD cases, only RA was significantly associated with a decreased BC risk  (data not shown). Multivariable regression models showed similar results . SimilarP=0.02). Results obtained from stratified models by year of case–control selection and unstratified models were similar, except for an attenuation of the risk deficit for ER-positive BC with RA in recent years  , whereas SLE was inversely associated only with ER-negative tumours  (data not shown). Multivariable analysis showed similar results . Heterog<0.0001) .Our study found a reduced BC risk associated with RA that did not differ by the ER-status of the tumour. We also observed a reduced risk in women with SLE that was confined to ER-negative tumours. None of the other SARDs investigated was associated with BC risk overall or by ER status.Our observed association for RA agrees with the results of a recent meta-analysis that reported a pooled standardised incidence ratio of 0.84 (95% CI=0.79–0.90) for breast cancer among women with RA (α) and transforming growth factor-beta (TGF-β) have been reported in patients with RA (α and TGF-β have been reported to suppress breast cancer cell proliferation, predominantly against ER-positive cells (The variation in BC risk with different SARD conditions might reflect differences in cytokine profiles. For example, high circulating levels of tumour necrosis factor-alpha (TNF- with RA , but notve cells . On the Possible differences in known BC risk factors that were not available in our study, such as parity, age at first birth, obesity, alcohol consumption, and use of hormone therapy, should also be considered. It seems unlikely that they completely explain the observed associations, as these are the risk factors for ER-positive, but not for ER-negative tumours . In addiOur study strengths include population-based sampling of the participants and a large sample size. The SEER–Medicare-linked database allowed the identification of incident breast cancer cases and provided detailed tumour information. In addition, we captured SARD conditions from inpatient, outpatient, and physician claims, minimising selection bias by capturing the whole spectrum of the diseases, including the one-quarter of SARD patients who were treated only as outpatients. Exclusion of the 1 year before case–control selection minimised the possibility of reverse causality and detection bias that could result from increased medical surveillance around the time of cancer diagnosis.Study limitations include the lack of information on important BC risk factors such as obesity and parity, as well as detailed medical information about the SARD conditions, including severity, duration, and detailed treatment history. We adjusted for the use of immunosuppressive medications; however, underestimation of the actual use is expected, especially for the less expensive oral medications, such as corticosteroids, because of the medication coverage rules of Medicare. The missing information on ER expression for one-fifth of the BC cases might be a source of bias. However, the equal distribution of SARD conditions in BC cases with known and unknown ER status suggests that substantial differential bias is unlikely.In conclusion, this study provides evidence for a reduced risk of BC in older women with RA and also suggests that older women with SLE may be at reduced risk for ER-negative BC, suggesting a possible role for systemic mediators of inflammation and immunity against BC cells.


Risk estimates for contralateral and ipsilateral invasive malignancies following age and histology specific in situ breast carcinomas were calculated using Poisson's regression analysis. The agreement between concordant and discordant morphologies of invasive and in situ breast cancer was measured using the kappa statistic. Women with in situ breast cancer showed a relative risk of 2.03 for contralateral and 3.94 for ipsilateral invasive breast cancer. The risk was higher for in situ carcinomas diagnosed before the age of 50 years and after lobular in situ breast cancers. A comparison of the risks during the past decades suggested that the risk of ipsilateral breast cancer has increased in Sweden but that of contralateral breast cancer has remained unchanged. In situ and the subsequent invasive breast cancers did not seem to share their morphologies.Sweden was the first country to establish a nationwide breast cancer screening service. We used the Swedish Family-Cancer Database to evaluate the risk of invasive carcinoma after  In Sweden, the incidence of DCIS has increased since the introduction of mammographic screening, representing now 15–20% of all breast malignancies, compared with 5% of the cases in prescreening era  after LCIS vary from 2 to 11 , lobular carcinoma (85202) and comedo carcinoma (85012). Breast cancer is often of mixed morphology at presentation and only the main type was recorded in the Swedish Cancer Registry , the date of immigration, the date of diagnosis of  program . The resin situ breast cancer. Kappa takes values between −1 and 1, where 0 indicates no determination and −1 or 1 would indicate that the morphology of the invasive carcinoma is completely determined by the histology of the prior in situ carcinoma. Values of the kappa statistic between 0.40 and 0.60 would suggest a moderate concordance.The kappa statistic was used as a measure of agreement between concordant and discordant morphologies of invasive and In situ ductal cancer increased from 7.2 to 8.7/100 000 (+3.1% per year) and, interestingly, the incidence of in situ lobular cancer remained constant (1.6/100 000). In situ comedo cancer showed a decreasing incidence rate from 1.7 to 0.8/100 000 (−11.1% per year).The incidence rate of invasive ductal cancer increased in the period 1993–2000 from 46.8 to 66.1/100 000 (+5.1% per year) and invasive lobular cancer showed an increase from 9.0 to 14.2/100 000 (+7.1% per year). A striking decrease was found for the comedo histology, with incidence rates from 6.1 to 1.1/100 000 (−18.6% per year). in situ breast carcinoma resulted in a two-fold increase in the risk of invasive cancer in the contralateral breast; the increase was particularly high if the in situ breast lesion was diagnosed before the age of 50 years. The risk increased with the time after diagnosis of the in situ carcinoma, although the increment was not statistically significant. The lobular histology of in situ breast carcinoma was associated with the highest risk for contralateral invasive breast cancer. The diagnosis of in situ carcinoma resulted in an RR of 3.94 for invasive ipsilateral cancer .The results of the Poisson regression analysis for contralateral breast cancer are presented in l cancer . Comedo in situ breast cancer slightly increased during the period 1991–1995 and it decreased thereafter, but were not statistically significant (in situ breast cancer in 1991–1995 and 1996–2000 were two times higher than in 1981–1990 (in situ carcinomas diagnosed at ages 50–60 years (results not shown).The risks of contralateral invasive cancer after nificant . In cont981–1990 ; the risin situ and invasive morphologies. Kappa values for all concordant and discordant histologies (results not shown for discordant histologies) were lower than 0.40, thus suggesting that the histology of the invasive cancers was not determined by the morphology of the preceding in situ carcinoma.in situ carcinoma in the literature (The treatment for the first breast cancer, the intense medical follow-up and the self-observation of the patient complicate the epidemiology of second breast cancers. Moreover, two breasts are at risk for the first cancer, but only one of them is at risk for ipsilateral or for contralateral breast cancer. Treatment, follow-up and definition differences have probably contributed to the incongruent results for invasive breast cancer after in situ breast carcinomas are premalignant, being possible precursors to invasive disease capable of metastasis (in situ carcinoma have resulted in better survival rates for patients and a higher incidence for invasive carcinoma. An advantage of the present study was the adjustment for age, family history and parity in the Poisson regression analyses. Unfortunately, information on possible confounders such as treatment received, use of contraceptives and stage of the cancer or tumour size was not available.All in situ carcinomas as DCIS. In contrast, the incidence of LCIS did not change. Earlier reports have shown increases in the incidence of DCIS and LCIS (The increased incidence of invasive ductal and lobular breast cancers found in the present study was similar to previous reports (In situ breast carcinoma diagnosed before the age of 50 years was associated with the highest risk of invasive cancer, the difference from other age groups was even statistically significant for ipsilateral breast cancer. In patients with mammographically detected ductal carcinoma in situ, treated with breast-conserving therapy, young patient age has been reported to be a risk factor for local recurrence (in situ cancer diagnosed at young ages (in situ carcinoma, but the number of cases analysed was small and the trend was not statistically significant. Our data agree with earlier results, that the highest risk of invasive cancer is reached after 42 to 60 months after diagnosis of in situ lesions. (in situ. Our results therefore disagree with the report suggesting that this risk was higher after DCIS than after LCIS (The risk of invasive cancer in the contralateral and ipsilateral breasts was higher after lobular than after ductal carcinoma in situ lesions was almost two times higher than the corresponding risk of contralateral breast cancer. This risk difference was only found after 1991, and was statistically significant for the period 1996–2000. Breast-conserving surgery became the treatment of choice for early breast cancer in Sweden during the 1980s, especially in areas with population-based screening (in situ breast carcinoma, for example, insufficient margin control, may have contributed to the observed risk increase (The estimated risk of ipsilateral invasive cancer after increase .in situ and subsequent invasive cancers share morphological features (The suggestion of previous studies that features .in situ carcinoma of the breast has increased in the last decade and that the introduction of screening and new treatments in Sweden seems to have modified the pattern of risk of invasive cancer after in situ carcinoma. The risk of invasive cancer in the ipsilateral breast has increased. Women with LCIS are at higher risk of invasive breast cancer than women with DCIS. The risks estimated in this study may help in clinical counselling.We conclude that the incidence of


Laboratory and epidemiologic studies have suggested a modifying effect of cardiac glycosides  on cancer risk. We explored the association between digoxin treatment and invasive breast cancer incidence among postmenopausal Danish women.We used Danish registries to identify 5,565 postmenopausal women diagnosed with incident invasive breast carcinoma between 1 January 1991 and 31 December 2007, and 55,650 matched population controls. Cardiac glycoside prescriptions were ascertained from county prescription registries. All subjects had at least 2 years of recorded prescription drug and medical history data. We estimated the odds ratio associating digoxin use with breast cancer in conditional logistic regression models adjusted for age, county of residence, and use of anticoagulants, non-steroidal anti-inflammatory drugs (NSAIDs), aspirin, and hormone replacement therapy. We also explored the impact of confounding by indication and detection bias.Digoxin was the sole cardiac glycoside prescribed to subjects during the study period. There were 324 breast cancer cases (5.8%) and 2,546 controls (4.6%) with a history of digoxin use at least 1 year before their index date (adjusted odds ratio (OR): 1.30; 95% confidence interval: 1.14 to 1.48). The breast cancer OR increased modestly with increasing duration of digoxin exposure . The association was robust to adjustment for age, receipt of hormone replacement therapy, coprescribed drugs, and confounding by indication. A comparison of screening mammography rates between cases and controls showed no evidence of detection bias.Our results suggest that digoxin treatment increases the risk of invasive breast cancer among postmenopausal women. Digitalis-derived cardenolides digitoxin and digoxin. These compounds exert their pharmacologic effect via inhibition of the Na+/K+ ATPase, which indirectly raises intracellular Ca2+ concentration, thus increasing the force of contractility in cardiac myocytes.Cardiac glycosides (CGs) are natural steroid toxins that have been used since the 18th century to treat congestive heart failure (CHF) and atrial fibrillation (AF) . Wom. Womet an = 142)  and appr+ ATPase ,6.et al. compared site-specific cancer incidence rates among digitalis-treated Norwegians patients with expected rates in the general population [Subsequent studies of the association between CG use and breast cancer incidence gave conflicting results. Haux pulation . Severalpulation . Also, Fpulation .Given the continued importance of CG medicines to treat heart disease and the inconsistent results from earlier studies of the association between this therapy and breast cancer occurrence, we examined the association between digoxin treatment and breast tumor incidence rate in a population-based prospective case-control study of postmenopausal Danish women.This study was approved by the Boston University Medical Campus Institutional Review Board and the Danish Registry Board.International Classification of Diseases, 8th revision  and 10th revision .This study was conducted within the female population of North Jutland and Aarhus Counties, Denmark . We usedControls were identified in the Danish Civil Registration System, which has tracked residential address, vital status, and date of emigration for the entire Danish population since 1968 . ControlWe used each subject's unique CPR number to link the case-control roster to county prescription databases ,13, whicWe identified cases of incident breast cancer in the hospital registries using ICD-8 and ICD-10 codes appropriate to the date ranges of the databases. ICD codes were also used to ascertain comorbid conditions for cases and controls  or never exposed (no record of digoxin prescription at least 1 year before the index date), and in finer terms according to the length of time between a woman's first digoxin prescription and her index date.a priori based upon established breast cancer risk factors that were also likely to influence receipt of digoxin. Age was initially controlled by matching cases to controls on year of birth. We also calculated each subject's exact age on her index date to adjust for residual confounding by age. We additionally considered confounding by coprescription of anticoagulants, non-steroidal anti-inflammatory drugs (NSAIDs), aspirin, and hormone replacement therapy (HRT). Anticoagulants are frequently prescribed for AF, and were associated with lower risk of urogenital cancer [Confounders were selected l cancer . NSAID ul cancer , and thel cancer . Aspirinl cancer , though l cancer . We therl cancer .Prescriptions for hormone replacement therapy were identified by ATC codes . Exposure to any of these drugs before the index date was classified as 'ever exposed to HRT' while exposure to none of them was classified as 'never exposed to HRT'. Similarly, we characterized ever/never exposure to anticoagulants, NSAIDs and aspirin by searching for ATC codes beginning with B01A, M01A, and B01AC06, respectively.We evaluated confounding by the medical indications for digoxin therapy by defining an alternative reference group of women who were never exposed to digoxin and who had a history of cardiovascular disease (excluding CHF or AF). We hypothesized these reference subjects should be more similar to the digoxin-treated women with regard to cumulative hormonal exposures and lifestyle factors that may modify risk for both heart disease and breast cancer. This reference group also facilitated evaluation of detection bias by allowing comparison of digoxin-exposed women to women with other serious histories who would likely have similar medical usage patterns. We further evaluated detection bias by comparing mammography usage rates between cases and controls. Dates of all mammography procedures among cases and controls were identified in hospital registries using appropriate Danish medical procedure codes. We analyzed mammography usage among women with index dates from 1 January 2006 onward, the period of our study when screening mammography would have been most common in Denmark. For each subject who had undergone mammography before her index date, we identified her most recent procedure and calculated the time elapsed between that procedure and the index date.We characterized the names, doses, and prescribing frequencies of the various digoxin products used over the study period. We computed the frequency and proportion of cases and controls by digoxin exposure status, prevalent medical conditions, use of other prescription drugs , and age on index date.We calculated the unadjusted odds ratios (ORs) and 95% confidence intervals (CIs) associating digoxin exposure categories with incident breast cancer and used conditional logistic regression to account for the matching factors and to adjust for exact age and past use of HRT, anticoagulants, NSAIDs, and aspirin. Due to the risk-set sampling design, the odds ratio approximates the incidence rate ratio associating digoxin exposure with incident breast cancer . All anaWe identified 5,565 cases and 55,650 matched population controls. Among the cases, 324 (5.8%) had ever had a digoxin prescription at least 1 year before her diagnosis date and 2,546 (4.6%) of controls had ever had a digoxin prescription at least 1 year before her index date. The distributions of cases and controls according to age, mammography usage, comorbidity and relevant prescription drug usage are shown in Table Table We observed a higher rate of breast cancer among ever-users of digoxin, relative to never users, in both unadjusted and adjusted analyses  [in vitro ER binding studies do not support this notion [Other previous research is consistent with our results ,22. Hauxs notion ,24.+/K+ ATPase in a variety of signal transduction pathways, with end effects in cell adhesion, survival, and proliferation [in vitro studies point toward a downstream antiproliferative effect of CGs but others leave open the possibility of cancer-promoting endpoints [+/K+ ATPase and the consequential effects appear to be highly dependent on the specific CG compound and the subunit makeup of the receiving ATPase [in vivo, consistent with our findings. With this study, we have isolated the association between a single cardiac glycoside, digoxin, and breast cancer incidence in a virtually unselected population of postmenopausal women.Recent laboratory findings implicate the Naferation . Severalndpoints . The intg ATPase ,26,27. TThe main strengths of this study are its large size, use of high-validity registry data to ascertain diagnoses, use of prospectively-recorded exposure information, and lack of selection in enumerating cases and controls.Our study design minimized the threat of selection bias, which can create the illusion of an exposure-disease association when, in fact, none exists . We had Our results are subject to distortion by residual confounding and misclassification of exposure and outcome. We took measures to address confounding by age, past exposure to other prescription drugs, and the medical indications for digoxin prescription. We saw little change in the unadjusted association after accounting for these factors. Digoxin is ordinarily prescribed at an age when most women no longer bear children, so it is unlikely that digoxin exposure is strongly associated with the well-characterized reproductive factors that affect breast cancer risk . We therWe were not able to adjust directly for body mass index (BMI), which is associated with both cardiovascular disease (CVD) and breast cancer . HoweverOur characterization of digoxin exposure was informed only by the number and strength of prescriptions filled by study participants; the prescription registry data did not permit calculation of actual daily doses taken by exposed subjects. Because prescription records were generated automatically before breast cancer diagnoses, we expect any exposure classification error to be non-differential in nature. We are not aware of published validation data on the classification of incident breast cancer in the hospital discharge registries. However, breast cancer diagnoses were recorded without express knowledge of exposure, so outcome misclassification is also expected to be non-differential. Since non-differential classification errors are expected to attenuate results, exposure and outcome misclassification cannot plausibly account for our positive association .We observed a modestly increased rate of breast cancer among postmenopausal women with any history of digoxin use, compared with women with no such use, after adjustment for age, use of other prescription drugs, and cardiovascular indications. The associations persisted in long-term exposure categories. While a number of laboratory studies of cardiac glycosides and female breast cancer have suggested protective effects, our results suggest that one specific cardiac glycoside, digoxin, moderately increases the incidence rate of breast cancer. This finding agrees with results from past studies; ,8,21 theAF: atrial fibrillation; BMI: body mass index; CG: cardiac glycoside; CHF: congestive heart failure; CI: confidence interval; CVD: cardiovascular disease; ER: estrogen receptor; HRT: hormone replacement therapy; ICD: International Classification of Diseases; OR: odds ratio; SIR: standardized incidence ratio.HTS reports receiving no fees, honoraria, grants or consultancies. The Department of Clinical Epidemiology is, however, involved in studies with funding from various pharmaceutical companies  as research grants to (and administered by) Aarhus University. None of these studies have relation to the present study. TPA, LAP and TLL declare no conflicts of interest.HTS, TPA and TLL conceived the study idea. TPA, TLL and HTS designed the study. HTS and LAP collected the data. TPA performed all data analyses, reviewed the literature and wrote the first draft of the manuscript. All authors edited the manuscript.