DNA methylation of CpG islands within the promoter region of genes is an epigenetic modification with an important role in the development of cancer and it is typically mediated by DNA methyltransferases (DNMTs). In cancer cells, global hypomethylation of the genome as a whole and regional hypermethylation of CpG islands have been reported. Four groups of DNMTs have been identified: DNMT1, DNMT2 (TRDMT1), DNMT3A and DNMT3B. DNMT2 uses the catalytic mechanism of DNMTs, but does in fact methylate RNA. Little is known about the significance of these genes in human breast cancer. In the study presented herein, we analyzed five distinct DNMT single SNPs with regard to potential associations with breast cancer risk.DNMT1 gene and two loci within the DNMT3A and DNMT3B genes, respectively. Statistics were calculated using the chi-squared and Fisher’s exact tests, and correlated with clinical parameters such as age, diagnosis, histology, TNM stage, hormonal receptor status, human epidermal growth factor receptor 2 (HER2) status, response to treatment and survival. Statistically significant results were obtained for correlations with the DNMT1 gene.In this study, we genotyped 221 female Caucasian breast cancer patients and 221 female Caucasian healthy controls, and we used five allele-specific real-time polymerase chain reaction (qPCR) assays. We selected one locus within the DNMT1, DNMT3A and DNMT3B genes were assessed. Statistical significance (P = 0.030) was identified for DNMT1 SNP : six women within the control group were GG homozygous (variant), while this mutation was absent in the breast cancer group.Five genomic loci within the DNMT1 SNP  GG homozygous genotype (variant) have a lower risk of developing breast cancer compared to heterozygous or wildtype genotypes. To date, alterations within the DNMT1 gene have not been reported to be associated with cancer in the Caucasian population.We conclude that women with the  BRCA1 and BRCA2 are genes associated with an increased risk of developing breast cancer. Inherited mutations in the p53 and PTEN genes have been observed in the context of familial syndromes, Li-Fraumeni syndrome and Cowden’s disease, which imply an increased risk of developing breast cancer. However, we need to consider that both Li-Fraumeni syndrome and Cowden’s disease are rare conditions[5C) transmission along with global DNA hypomethylation[Breast cancer is the most frequent malignancy in women and it is also the leading cause of death among women aged 40 to 50 years. Numerous risk factors of breast cancer have been reported in the literature and the most relevant factors are geographical variations, lifestyle, age at the time of diagnosis, age at first pregnancy, age at menarche, age at menopause and family history. hylation. Accordihylation.DNA methylation plays an important role in the control of gene expression in mammalian cells. In breade novo and the maintenance of methylation , DNMT3A and DNMT3B-12. DNMTde novo methylation and they are important for the generation of methylation patterns during embryogenesis[De novo DNMTs methylate cytosine to m5C post-replicatively from unmethylated DNA[Asp cytosine in position 38 (C38)[DNMT3A and DNMT3B are mainly involved in ogenesis. De novoated DNA. The two38 (C38). It is t38 (C38).DNMT1, DNMT3A and DNMT3B SNPs on DNMT gene expression and associated enzymatic activities, a number of gene loci were identified in silico on the basis of the information obtained from literature and SNP databases: DNMT1 SNP , DNMT3A SNPs  and DNMT3B SNPs . In this context, the DNMT2 (TRDMT1) gene has not been considered, because human DNMT2 is an RNA methyltransferase and this study was primarily focused on genes coding for DNMTs, and because they have been reported to play an important role in the development of cancer[In order to assess the relevance of DNMT SNPs and the risk of developing breast cancer.With the study presented herein, it was our intention to assess whether there was a correlation between In this study, we analyzed 221 DNA samples isolated from PBMCs of female Caucasian breast cancer patients and 221 DNA samples from female Caucasian healthy donors. Patient samples were collected at the Department of Obstetrics and Gynecology, Saarland University Medical Center, Germany, between 2001 and 2010. Control samples from healthy donors were collected at the Central Laboratory, University of Heidelberg Medical Center, Germany. All patients in the control group were below the age of 30 years. Genomic DNA was isolated from PBMCs and whole blood samples using the NucleoSpin Blood kit , in accordance with the manufacturer’s instructions. Analyses of DNA purity and quantification were performed using a NanoDrop spectrophotometer .DNMT1, DNMT3A and DNMT3B genes via in silico analyses using data from the National Center for Biotechnology Information , Applied Biosystems , GeneCards and HapMap. These SNPs carried the potential to affect the qualitative and quantitative expression of the DNMTs mentioned, and to search for potential associations with clinical parameters of breast cancer patients , DNMT3A SNPs  and DNMT3B SNPs . DNMT3B SNP  was selected on the basis of the publication by Cebrian et al.[DNMT3B gene. The DNMT1 SNP  was recently analyzed in Chinese women by Sun et al.[We identified SNPs within the coding and promoter regions of the n et al. and theyn et al.. The othAll samples of the selected tag SNPs were genotyped by real-time polymerase chain reaction (qPCR) or on TaqMan probes . qPCRs were carried out on 10 ng of genomic DNA using TaqMan universal PCR Master Mix . Forward and reverse primers were labeled VIC and FAM (designed by Applied Biosystems) in a 5 μl reaction Table . The amphttp://www.tufts.edu/~mcourt01/Documents/Court%20lab%20-%20HW%20calculator.xls), by comparing the observed and expected genotype frequencies for cases and controls. Genotyping data generated for patients and controls were correlated with clinical parameters  status, human epidermal growth factor receptor 2 (HER2) status, response to chemotherapy and survival) on the basis of the chi-squared and Fisher’s exact tests. P values ≤0.05 were considered to be statistically significant. All analyses were performed using the statistical analysis software SPSS Statistics Version 17 for Windows .The Hardy-Weinberg equilibrium was tested using the Simple Hardy-Weinberg Calculator by Michael H Court  and DNMT3B SNP . All women were homozygous for the DNMT3A GG and DNMT3B GG genotype. Based on information obtained from the NCBI database, the frequency of this polymorphism within the Caucasian population was reported to be 1.4% for the DNMT3A SNP  on the basis of 74 samples, 1.4% for the DNMT3B SNP  on the basis of 74 samples and 1.5% for the DNMT3A SNP  on the basis of 68 samples  revealed no statistical significance. The correlation of the genotyping results between cases and controls showed no coherence for the DNMT3B SNP  was analyzed earlier as reported in the literature[DNMT3B SNP and corresponding clinical parameters. Our genotyping data comparing patients and controls varied between 1.4% and 2.3%  in central European Caucasian females) , which appears to protect women against developing breast cancer.We identified one additional statistically significant correlation for P value of our results was 0.030, in accordance with Fisher’s exact test. The genotyping results showed that within the control group, 6 out of 221 women were GG homozygous (variant), while the homozygous variant genotype was not found in the group of breast cancer patients  and DNMT1 SNP ) reached statistical significance . For DNMT3A, we assessed two SNPs , which revealed the homozygous wildtype allele in all samples that were studied (cases and controls). DNMT3A SNPs with an impact on protein levels would have been of special interest, since Ley et al. showed that DNMT3A mutations are highly recurrent in patients with de novo acute myelogenous leukemia (AML) and with an intermediate-risk cytogenetic profile, and these patients also had a significantly reduced survival[de novo MDS and their association with secondary AML[de novo DNA methylation and has been shown to be over-expressed in numerous types of cancer[Two out of five SNPs in the survival. Furtherdary AML. These mdary AML. AML andf cancer,28-31, tf cancer,33. The f cancer,28-31.DNMT3B SNPs, of which DNMT3B SNP  failed to be confirmed in the female Caucasian population, showing only the wildtype genotype in all samples. DNMT3B SNP  showed a significant difference among cases and controls  with the T allele (variant) associated with a reduced risk of developing breast cancer, while the statistical analysis of the correlation with clinical parameters was not significant. Montgomery et al. described a strong association between the C allele of the DNMT3B promoter polymorphism  and the risk of developing breast cancer, compared to TT homozygotes[In this study, we assessed two interesting ozygotes.DNMT1 and DNMT3B genes, and the DNMT3B SNP , which was also analyzed in our study[DNMT3B SNP , were rejected from the analysis because either the minor allele frequencies were less than 0.005 in the study population or they were found to deviate from the Hardy-Weinberg equilibrium in controls. Thirteen SNPs in the DNMT3B gene  were included in the analysis, but Ye et al. observed no association between these SNPs and breast cancer among Chinese women[DNMT3B SNP  (odds ratio (OR) CT versus CC = 0.85 (0.77 to 0.94); OR TT versus CC = 0.89 (0.64 to 1.23); P heterogeneity = 0.007; P trend = 0.003). This association was in accordance with our statistical analysis, but Ye et al. could not confirm an apparent association of the DNMT3B polymorphism  with the risk of breast cancer among Chinese women, since this SNP was found to have a minor allele frequency of less than 0.05 in the study population and was excluded from the analysis[DNMT3B gene  is only relevant in the female Caucasian population and not among the female Chinese population, which might be due to the difference of the allele frequency of this mutation between Caucasian (12.5%) and Chinese patients (1.2%), according to the NCBI database (since genetic mutations often vary between ethnic groups).Ye et al. genotyped a total of 47 SNPs in the ur study. Twenty-se women, which wanalysis. ConsequDNMT1 SNP  which could be relevant as a risk of developing breast cancer in the female Caucasian population. In relation to breast cancer risk, DNMT1 gene polymorphisms have been solely reported in association with sporadic infiltrating ductal breast cancer among Chinese women[DNMT1 gene , and reported significance between the SNPs and the PR and p53 status (p53-positive disease)[DNMT1 gene , but they did not find any apparent association of DNMT1 mutation with the risk of breast cancer among Chinese women[DNMT1 gene[DNMT1 SNP rs2228612, the frequency of the GG genotype (variant) of rs2228612 was higher in cases compared to controls (22.5% versus 14.5%) (OR AG versus GG = 1.71 (1.06 to 2.78); OR GG versus AA 1.75 (1.13 to 2.72); P heterogeneity = 0.044; P trend = 0.013)[DNMT1 is a maintenance methyltransferase which attaches methyl groups to hemimethylated DNA during replication. To date, no genetic polymorphisms have been reported for DNMT1 in association with breast cancer in the female Caucasian population. In the se women. Xiang edisease). Ye et ase women. A recenDNMT1 gene polymorphism  and the risk of breast cancer in the female Caucasian population. The correlation of cases and controls showed a statistical significance with a P value of 0.030 (Fisher’s exact test), whereas the statistical analyses of frequency differences between cases and clinical parameters revealed no significant results. While six female controls were GG homozygous (variant), there were no females in the group of breast cancer patients with the GG homozygous genotype (variant). Therefore, our results show that the DNMT1 gene polymorphism  with the g-allele (variant) is associated with a reduced risk of developing breast cancer.In this study, we were able to show a significant association between the DNMT3A SNPs  and the DNMT3B SNP  do not exist in the female Caucasian population. Otherwise, we could demonstrate a statistical significance in the differences of allele frequencies of cases and controls of the DNMT3B SNP , like in the study of Cebrian et al.[In conclusion, our results showed that the two n et al..DNMT1 SNP  variant in the correlation of cases and controls. Women with the G allele (variant) show a lower risk of developing breast cancer. The diagnostic impact of this new marker needs to be validated in further clinical studies in larger populations and within different ethnic groups, but it presents a further step towards a more individualized diagnosis.Finally, we identified a significant association of the AML: Acute myelogenous leukemia; BMI: Body mass index; C: Cytosine; CML: Chronic myelogenous leukemia; DNMT: DNA methyltransferase; ER: Estrogen receptor; HER2: Human epidermal growth factor receptor 2; m5C: 5-methylcytosine; MDS: Myelodysplastic syndrome; NCBI: National Center for Biotechnology Information; OR: Odds ratio; PBMC: Peripheral blood mononuclear cell; PR: Progesterone receptor; qPCR: Real-time polymerase chain reaction; SNP: Single nucleotide polymorphism; TNM: TNM Classification of Malignant Tumours.All authors declare that they have no conflict of interest in association with this study.KK carried out the molecular genetic studies and performed the statistical analysis with MFO. UM planned the study. KK and UM worked on the manuscript. All authors read and approved the final manuscript.Table S2. Explanation of the clinical parameters. The age of the patients are defined as age at the time of recruitment. Data of breast cancer patients. Click here for file


Evidence of the accuracy of self-reported mammography use among women with familial breast cancer risk is limited. This study examined the accuracy of self-reported screening mammography dates in a cohort of 1,114 female relatives of breast cancer cases, aged 26 to 73 from the Ontario site of the Breast Cancer Family Registry. Self-reported dates were compared to dates abstracted from imaging reports. Associations between inaccurate recall and subject characteristics were assessed using multinomial regression. Almost all women  accurately reported their mammogram use within the previous 12 months. Women at low familial risk , who reported 1 or fewer annual visits to a health professional , exhibited a lower perceived breast cancer risk , and reported a mammogram date more than 12 months previous , were significantly more likely to inaccurately recall their mammogram date. Women with varying levels of familial risk are accurate reporters of their mammogram use. These results present the first evidence of self-reported mammography recall accuracy among women with varying levels of familial risk. Having a family history of breast cancer has been established as one of the most important risk factors for the development of breast cancer –3. A redSelf-reported data is often used in epidemiologic research evaluating the use of cancer screening. The validity of self-reported mammography use has been studied extensively in women with population-level breast cancer risk –18. OverFew studies have validated self-reported mammogram data among women with familial risk , 26. TheFamily History Questionnaire (FHQ). Respondents meeting a defined set of genetic risk criteria and a random sample (25%) of those not meeting criteria were asked to participate in the Ontario site of the BCFR. Of the 2,587 eligible women, 1,851 (71.5%) probands participated. Probands were then asked for permission to contact specific living relatives . An invitation letter to participate in the Ontario site of the BCFR was sent to all relatives , and the 5,122 who agreed to be contacted were mailed an Epidemiology Questionnaire (EQ) from 1998 to 2004. This study utilized data from a cohort of female relatives of incident cases of invasive breast cancer identified from the Ontario site of the Breast Cancer Family Registry (BCFR) funded by the United States National Cancer Institute. Details of the BCFR and the Ontario site have been previously described , 28. BriPersonal History and Screening Questionnaires (PHSQ) were administered to update changes in demographic characteristics, cancer screening behaviors and breast outcomes. All female relatives enrolled in the Ontario site of the BCFR, who completed an EQ and were 20 to 69 years of age and unaffected by breast cancer at the time of the proband's diagnosis, were eligible to participate. From the 3,374 participating female relatives, 1,514 met all study criteria. A baseline PHSQ was sent between 2005 and 2007 to the 1,514 eligible women. Of these women, 1,114 (73.6%) completed the baseline PHSQ. A year 1 PHSQ was sent to 1,077 eligible women approximately one year following the baseline PHSQ and 975 (90.5%) completed it. A year 2 PHSQ was sent to 969 eligible women and 882 (91.0%) completed it.This prospective cohort study was conducted several years after initial recruitment of relatives to the Ontario site of the BCFR. Baseline, year 1, and year 2 n = 1114), year 1 (n = 975), and year 2 (n = 882) PHSQs and reported having a mammogram since their previous interview  were eligible. Women were excluded if they had a personal breast cancer diagnosis, they did not have a first-degree family history of breast or ovarian cancer, they did not provide consent to release the imaging report, the imaging report was not available from the imaging centre, or the indication for the mammogram was for nonscreening purposes. Final sample sizes included 699 women at baseline, 469 at year 1 and 456 at year 2. This study was approved by the Research Ethics Boards of Mount Sinai Hospital, the University Health Network, and University of Toronto.  Women who completed baseline (EQ) was self-administered during recruitment of female relatives to the Ontario site of the BCFR and collected detailed information on demographics and key behavioral risk factors for breast and ovarian cancer. As several years had elapsed since completion of the EQ, three subsequent questionnaires  of similar contents were telephone-administered to update changes in key demographic and health behavior characteristics and collect detailed information on breast cancer screening. Eligible participants were sent an introductory letter with a copy of the PHSQ approximately two weeks prior to being contacted by telephone. This allowed participants to recall dates and events and allowed reference to the questionnaire during the interview. This method was found to result in higher response rates and more complete data than achieved by self-administered questionnaires [Information was obtained from four questionnaires. The first  or their age at last mammogram. Women were also asked to indicate if the main reason for the last exam was for screening , or nonscreening . For women who provided consent, imaging reports were obtained from the imaging facility and abstracted for: (i) imaging date ; (ii) indication (screening or diagnostic); and (iii) breast imaging-reporting and data system (BI-RADS) classification [Self-reported dates and reasons for mammograms were obtained from fication . FHQ completed by the proband, using a modified definition of previously referenced groups for familial breast cancer risk [PHSQ. Use of clinical breast examination (CBE) and BRCA1/2 genetic testing was updated at each PHSQ and based on self-reported use since the previous interview. Perceived risk of developing breast cancer  was updated at each PHSQ and determined using a question adopted from Lipkus et al. [Classifications of familial breast cancer risk were based on data collected by the cancer cer risk , 33.Tabls et al. . Women wPHSQ were summarized. Self-reported screening mammogram dates and abstracted imaging dates (“gold standard”) were compared to assess rates of agreement. Inaccurate recall was classified as overestimation or underestimation of the time since last mammogram. The difference in months between self-reported and abstracted mammogram dates was compared among women who inaccurately reported mammogram dates. Multinomial logistic regression was used to estimate adjusted associations between inaccurate recall and a number of sociodemographic, health behavior and cancer screening characteristics. All models included familial breast cancer risk, age at interview, and number of days since last mammogram (except for models including time since last mammogram in categories). As women are clustered within families, robust variance estimation was used to estimate all confidence intervals (CI) [P values at a 5% testing level. Distributions of sample characteristics at each als (CI) , 36. AllEQ. Most women (61.5%) perceived their lifetime breast cancer risk to be higher than average. Since completion of the EQ, 34.3% reported a mammogram within the previous 6 months, 33.1% reported a mammogram 7 to 12 months prior, and 32.6% reported that their last mammogram was more than 12 months prior. Distributions of sociodemographic and health behavior characteristics were similar at years 1 and 2. At baseline, 42.6% of women were classified as low familial risk, while 26.2% and 31.2% were moderate and high risk, respectively . The majEQ and PHSQ) was longer (5 to 7 years). Among women who inaccurately recalled the date of their last mammogram, proportions of women who underestimated and overestimated this time interval were approximately equal at baseline and year 1, but more women overestimated at year 2. Women who underestimated this interval did so by a longer amount of time at baseline and year 2, compared with women who overestimated . This difference was statistically significant at baseline (P = 0.036). Most women were able to report the month and year  of their last mammogram, while 24.2% at baseline, 11.1% at year 1 and 6.6% at year 2 were only to report the year or their age . Among wAmong women who reported the month and year of their last mammogram, most  reported it was within the previous 12 months . This waAt baseline, women with low familial risk were more likely to overestimate the time since last mammography compared to high-risk women . Women who reported an average of 1 or fewer visits to a health professional per year were almost twice as likely to overestimate the time since last mammogram  compared to women with 2 to 3 visits per year, as were women who perceived their breast cancer risk to be below or the same as average  . Women wOur study presents some of the first evidence of the accuracy of self-reported screening mammography use among women with varying levels of familial risk. We found that self-reported screening mammogram use within a 12-month period is highly accurate (over 95%). Additionally, 62% of women at baseline, 79% at year 1, and 80% at year 2 accurately reported the exact timing (month and year) of their last screening mammogram. While we did not find systematic evidence of telescoping, we found that the difference in months between self-reported and abstracted dates was significantly larger in women who telescoped the date at baseline. Fewer visits to a health professional per year, lower perceived breast cancer risk, and having a mammogram more than 12 months prior were significantly associated with inaccurate recall, while the association between inaccurate recall and level of familial risk approached significance. The high rates of recall accuracy we observed differ from previous studies, most of which focused on women in the general population. For example, two studies , 37 repoTelescoping has been one of the most consistent findings of previous studies –23. PijpStudies which have previously examined predictors of mammogram recall have reported inconsistent results, likely due to differences in study populations and definitions of accurate recall. When examining predictors of recall, most previous studies have not further classified inaccurate recall as overestimation or underestimation of the time since last mammogram. Pijpe et al. , who didn = 3 at baseline, n = 22 at year 1, n = 16 at year 2). As women were mailed a copy of the questionnaire approximately two weeks before being contacted by study staff, they could have checked their personal records or otherwise verified the date of their last mammogram prior to completing the questionnaire. This may have inflated rates of recall accuracy. Finally, it should be noted that the accuracy of recall may differ in populations with different breast cancer screening recommendations and systems of healthcare. Our study is unique in that it is one of the first to examine the accuracy of self-reported mammography data among a population of women with a range of family histories of breast cancer. The population-based recruitment, large sample size, longer follow-up period, and exclusion of diagnostic mammograms are also notable strengths. This study has several limitations. Due to the challenge of verifying reports from multiple service providers, this study was not designed to assess recall among women who reported nonuse of mammography. Previous evidence suggests women are unlikely to falsely report the nonuse of mammography , 22, 23.Self-reported mammography use data is widely relied upon in epidemiologic research for measuring adherence to breast cancer screening guidelines. Overall, our study found women with familial breast cancer risk to be extremely accurate reporters of their mammogram use within 12 months and reasonably reliable reporters of the exact timing of their mammogram attendance. Recall was poorer among women with low familial breast cancer risk and lower perceived breast cancer risk, and when asked to recall screening episodes more than one year prior. Where possible, strategies demonstrated to improve the accuracy of self-reported data should be employed by researchers . Women w


Advances in genetic technologies are expected to make population-wide genetic testing feasible. This could provide a basis for risk stratified cancer screening; but acceptability in the target populations has not been explored.We assessed attitudes to risk-stratified ovarian cancer (OC) screening based on prior genetic risk assessment using a survey design. Home-based interviews were carried out by the UK Office of National Statistics in a population-based sample of 1095 women aged 18–74. Demographic and personal correlates of attitudes to risk-stratified OC screening based on prior genetic risk assessment were determined using univariate analyses and adjusted logistic regression models.n = 61). Most women (85 %) said they would ‘probably’ or ‘definitely’ take up OC genetic testing; which increased to 88 % if the test also informed about breast cancer risk. Almost all women (92 %) thought they would ‘probably’ or ‘definitely’ participate in risk-stratified OC screening. In multivariate logistic regression models, university level education was associated with lower anticipated uptake of genetic testing (p = 0.009), but with more positive attitudes toward risk-stratified screening (p <0.001). Perceived risk was not significantly associated with any of the outcome variables.Full data on the key analytic questions were available for 829 respondents . Relatively few respondents felt they were at ‘higher’ or ‘much higher’ risk of OC than other women of their age group (7.4 %, These findings give confidence in taking forward research on integration of novel genomic technologies into mainstream healthcare. Although ovarian cancer is one of the leading causes of cancer death in women , populatCurrently, efforts are under way to develop and validate models for risk stratification, early detection and diagnosis of OC which incorporate clinical, epidemiological, proteomic and genetic data . Genetic models will include high penetrance genes associated with hereditary breast and ovarian cancer, BRCA1 and BRCA2, as well as lower penetrance genes associated with ovarian cancer susceptibility BRIP1, RAD51C and RAD51D –9. GivenHowever, the success of any cancer screening programme depends on its acceptability in the target population . Given tResearch investigating the predictors of interest in cancer genetic testing has predominantly focused on individuals with a strong family history of cancer. In these studies, age, ethnicity and perceived risk have frequently been shown to be associated with higher interest in testing . HoweverThe aims of the present study were therefore to explore public attitudes of women in the UK towards risk-stratified screening based on prior genetic risk assessment for OC, and identify demographic and personal predictors of anticipated uptake of both genetic testing and screening.n = 2010) are identified from the Royal Mail’s Postcode Address File using stratified random probability sampling. The selected addresses are contacted up to eight times at different times and days of the week to maximize response rates. One person aged over 16 from each household is randomly chosen to complete a computer-assisted, face-to-face interview with a trained interviewer. Verbal informed consent was obtained from participants before commencing the survey. Since this study was a population-based anonymous survey, it was exempt from ethical approval, as per guidelines set out by the UCL Ethics committee for non-NHS research (http://ethics.grad.ucl.ac.uk/exemptions.php).Data were collected as a part of the Office of National Statistics (ONS) ‘Opinions and Lifestyle’ survey in January and March 2014. This is a population-based survey of adults aged 16–94 years conducted by the ONS on behalf of government departments, non-governmental agencies, and academic institutions. The sample covers Great Britain , excluding the Isles of Scilly and the Scottish Highlands and Islands. Modules include a range of topics; minimizing the risk of respondent bias for specific topics. Respondents are given the option to withdraw at any time. Each month, households . This produced a total score of 68 for all questions, which translates to the reading level expected in grade 6 (age 12).The attitude items were specifically developed for this survey. The questions were read aloud, but we used the Flesch Reading Ease formula and the Flesch-Kincaid Grade Level formula to provide a formal estimate of comprehension . The NHS is the provider of healthcare in the UK. The NHS runs all the organised cancer screening programmes. Response options for all questions were: ‘If you were invited for NHS screening for ovarian cancer, would you take up the offer’ ?‘It’s possible that women who are found to be at higher genetic risk will be offered more frequent screening for ovarian cancer, and those found to be at lower risk will be offered less frequent screening. What do you think of the idea of varying the frequency of ovarian screening’ ?Two questions addressed attitudes to risk-stratified ovarian cancer screening:Compared with other women of your age, what do you think are your chances of getting ovarian cancer’? Response options were: ‘much lower than others’, ‘lower than others’, ‘the same as others’, ‘higher than others’, ‘much higher than others’. For the current analyses, the first three and last two categories were combined to reflect lower/average vs. higher perceived risk.Perceived relative risk of ovarian cancer was assessed with one question: ‘http://www.ons.gov.uk/ons/index.html). Age was included as a continuous variable in statistical analyses. Ethnicity was classified as ‘White’ vs. ‘Ethnic minority’ because the individual ethnic minority sub-groups were small; making subdivision inappropriate. Educational attainment was classified as university degree or equivalent vs. below degree level. Marital status was coded as married/cohabiting vs. single/widowed/divorced.Demographic data were provided by ONS from their standard survey items  the three items on anticipated uptake of genetic testing, and ii) the two items on attitudes to risk-stratified screening. The first model included only demographic characteristics, and the second one added perceived risk. Bonferroni corrections were used to correct for multiple testing, so α was set at 0.05/5 = 0.01.Statistical analyses were carried out using the Statistical Package for Social Sciences (SPSS) version 20.0 . Univariate χn = 166) of the 2010 selected households were not eligible because they were businesses or empty properties. Of 1844 eligible households, 9 % (n = 171) could not be contacted and 33 % (n = 608) declined to take part in the ONS survey. In the March wave, 8 % (165) of selected households were not eligible because they were business or empty properties. Of the 1853 eligible households, 13 % (n = 237) could not be contacted and 31 % (n = 578) chose not to take part. Therefore, the overall response rate was 57 %, comparable to previous ONS surveys (http://tinyurl.com/z8apjt6). The total female sample was n = 1095.Data for the current study were collected in two waves: January and March 2014. In the January wave, 8 % (n = 138) or missing data (n = 128), the final sample for analysis consisted of 829 women aged 18–74 . Most were ‘White’ , just over half were married or cohabiting , and just over a quarter  were at least university educated. With the exception of education, which was higher in this sample, demographic characteristics were comparable to the UK female population in this age group (ONS Census 2011: http://www.ons.gov.uk/ons/guide-method/census/2011/uk-census/uk-census-data-releases/index.html).After exclusion of individuals who were ineligible due to age (n = 734), which increased to 88.3 % (n = 819) if the test also informed about breast cancer risk. Anticipated ‘probable’ or ‘definite’ uptake remained high  if the test result indicated they were at very high risk and the appropriate risk management recommendation was risk-reducing surgery.Overall, interest in genetic testing for ovarian cancer was very high. Most women said they would ‘probably’ or ‘definitely’ take up an offer of genetic testing . University level education was not associated with anticipated uptake of genetic testing in univariate analyses, but it became significantly associated with lower anticipated uptake once demographic and personal factors were taken into account .There were few demographic differences in anticipated uptake of genetic testing for ovarian cancer risk Table . Ethnic n = 61). Perceived risk was not significantly associated with anticipated uptake of ovarian cancer genetic testing in univariate analyses (p = 0.729), and adding it to the multivariate model made no difference to the other associations (data not shown).Relatively few participants felt that they were at ‘higher’ or ‘much higher’ risk of ovarian cancer than other women of their age group .p <0.001).There were few demographic associations with anticipated uptake of ovarian screening or attitudes to risk-stratified screening. University level education was associated with more positive attitudes towards risk-stratified screening in univariate and multivariate analyses . Slightly more respondents who perceived themselves as higher risk than lower risk supported risk-stratified screening (69.8 % vs. 67.3 %), but the difference was not significant (p = 0.685). Perceived risk was also not significantly associated with any of outcomes in multivariate models (data not shown).Perceived risk was not significantly associated with anticipated uptake of ovarian cancer screening in univariate analyses  which reduces the likelihood of response bias related to the specific subject matter. The sample had been recruited by stratified random sampling which is the ‘gold-standard’ of survey recruitment; therefore, it is broadly representative of the UK population of women aged 18–74. We adjusted analyses for multiple testing, which gives confidence in the robustness of the current findings.The study also had limitations. At this stage, neither population-based genetic testing nor OC screening is recommended. Therefore, all the questions alluded to hypothetical scenarios rather than real testing/screening offers, and these have been found to correlate only modestly with actual behaviour , 26. HowStratified cancer screening based on genetic risk is considerably more complex than simple age-stratified approaches and raises many ethical, legal, social and organisational aspects . ChowdhuThe findings from this study suggest that women in the UK are ready to accept the introduction of population-based genetic testing for personalisation of future cancer screening programmes; giving confidence in taking forward research on integration of novel genomic technologies into mainstream healthcare.


In vitro and animal data suggest that cadmium, a heavy metal that contaminates some foods and tobacco plants, is an estrogenic endocrine disruptor. Elevated estrogen exposure is associated with breast, endometrial, and ovarian cancer risk.Background: Objectives: We examined the association between dietary cadmium intake and risk of these cancers in the large, well-characterized Women’s Health Initiative (WHI).Methods: A total of 155,069 postmenopausal women, 50–79 years of age, who were enrolled in the WHI clinical trials or observational study, participated in this study. We estimated dietary cadmium consumption by combining baseline food frequency questionnaire responses with U.S. Food and Drug Administration data on food cadmium content. Participants reported incident invasive breast, endometrial, or ovarian cancer, and WHI centrally adjudicated all cases through August 2009. We applied Cox regression to estimate adjusted hazard ratios (HRs) and 95% CIs for each cancer, comparing quintiles of energy-adjusted dietary cadmium intake.Results: Over an average of 10.5 years, 6,658 invasive breast cancers, 1,198 endometrial cancers, and 735 ovarian cancers were reported. We observed no statistically significant associations between dietary cadmium and risk of any of these cancers after adjustment for potential confounders including total dietary energy intake. Results did not differ in any subgroup of women examined.Conclusions: We found little evidence that dietary cadmium is a risk factor for breast, endometrial, or ovarian cancers in postmenopausal women. Misclassification in dietary cadmium assessment may have attenuated observed associations.http://dx.doi.org/10.1289/ehp.1307054Citation: Adams SV, Quraishi SM, Shafer MM, Passarelli MN, Freney EP, Chlebowski RT, Luo J, Meliker JR, Mu L, Neuhouser ML, Newcomb PA. 2014. Dietary cadmium exposure and risk of breast, endometrial, and ovarian cancer in the Women’s Health Initiative. Environ Health Perspect 122:594–600;  Cadmium, a carcinogenic heavy metal, is released into the environment as a result of industrial and agricultural activities . Tobaccoin vitro .Study population. We selected study participants from the WHI, a large longitudinal study of postmenopausal women, 50–79 years of age, comprising observational study (OS) and randomized clinical trial (CT) arms. Details of the study design and recruitment have been extensively described ], medical history, lifestyle , and physical activity , specifically designed for WHI and previously described in detail . These FTo estimate dietary cadmium intake, we adapted methodology commonly used for dietary micronutrient estimates  and descn = 57) in our calculation of dietary cadmium.The arithmetic mean of cadmium content (milligrams per 100 g prepared weight) of all available samples of each food was assigned as the cadmium content for that food. Thus, we averaged over year and region of collection to estimate each food’s cadmium content. To investigate whether this approach ignored important regional or secular trends in the cadmium content in foods, we examined measurements from the 20 foods with the highest reported cadmium concentrations. Regional variation was < 20% of the overall mean for 13 of the 20 top cadmium-containing foods; maximum variation of < 40% was observed for lettuces. No region with systematically higher or lower cadmium values was identified. Year-to-year variation in the 20 foods with highest mean cadmium concentration reached 50% of the mean cadmium. Qualitatively, however, trends with year of measurement were observed for only three of these 20 foods . Because of limited variation by region and year of measurement, we opted to include all available data from the FDA TDS for each food in order to obtain the best estimate of mean cadmium content of food typically consumed in the United States. We did not use baby foods  below the LOD were retained in analysis.We matched each of 302 food and beverage components comprising the 120 FFQ line items on the WHI FFQ to one of the foods analyzed by the FDA, based on the food names. To allow inclusion of participants at the Hawaii clinical center, 27 additional component foods specific to the Hawaii FFQ were matched. For component foods for which no obviously similar food was analyzed by the FDA, we relied on food “mapping” created by the FDA for the TDS (FDA 2013). In summary, 154 component foods were direct matches based on the food names; 122 component foods were close but differed in either unspecified details  or preparation ; 26 component foods were matched using the TDS mapping file . Therefore, we attributed a cadmium content to every component food.http://www.ncc.umn.edu/products/ndsr.html). Nutrient and cadmium calculations were performed by the Nutrition Assessment Shared Resource of the Fred Hutchinson Cancer Research Center .The FFQ analytic program calculated average annual servings of each FFQ line item, adjusted to sex-specific portion sizes, and estimated nutrient intakes based on the University of Minnesota Nutrition Data System for Research  of urine samples was measured at the Trace Elements Research Laboratory at the Wisconsin State Laboratory of Hygiene , using sector field inductively coupled plasma mass spectrometry  as described  FFQ data  or without follow-up information for cancer diagnosis (n = 650). We also excluded women with a previous diagnosis of a cancer of interest  from analysis of that specific cancer. For analyses of endometrial cancer risk, we excluded women with hysterectomy before enrollment . For analyses of ovarian cancer risk, we excluded women reporting bilateral oophorectomy before enrollment . We included women with missing information on a given variable as a separate category for adjustment; 10.4% of participants had missing information on one or more variables. A total of 155,069 women were included in one or more analyses .Follow-up for cancer and censoring. Participants updated medical history annually (OS) or semiannually (CT) through a mailed self-administered or telephone-administered questionnaire. Breast, endometrial, and ovarian cancers reported by participants were adjudicated by WHI Clinical Coordinating Center (WHI-CCC) staff and physician review of medical records  and were therefore not included as outcomes but did contribute time-at-risk before in situ diagnosis. For endometrial cancer analyses, women were censored at hysterectomy . Hysterectomy was reported on annual or semiannual updates and adjudicated by WHI-CCC for hormone CT participants; analysis of OS participants relied on self-reported hysterectomy information. No updated information on oophorectomy was collected during the study, to enable censoring in analysis of ovarian cancer risk. Death was ascertained through clinical center follow-up of family reports and routine checks with the National Death Index. records . SeparatStatistical analyses. Multivariable Cox proportional hazards regression was applied to estimate adjusted cancer hazard ratios (HRs) with 95% CIs by quintile of dietary cadmium, adjusted for total energy intake by the residual method , alcohol consumption , combined estrogen plus progesterone hormone therapy , unopposed estrogen hormone therapy , age at first birth , age at menarche , age at menopause , physical activity , and cigarette smoking history . Breast cancer analyses were additionally adjusted for mammography in the 2 years before baseline (yes/no). In further analyses, we additionally adjusted for daily medium-sized servings of vegetables , and daily medium-sized servings of grains (quartiles). To investigate alternatives to vegetable servings and grains, we performed additional analyses adjusted for intake of zinc and iron from diet and supplements (quartiles) along with servings of vegetables and servings of grains; or replaced adjustment for servings of vegetables and grains with computed grams of fiber and carbohydrates consumed daily (quartiles). Finally we examined adjustment for pack-years of smoking, in addition to smoking status.We selected confounders, each measured at baseline, based on knowledge of risk factors for breast, endometrial, and ovarian cancer, and sources of cadmium exposure. Multivariable models were stratified for enrollment age in bands , and on WHI component (OS or CT), and adjusted for age (years) (in addition to stratification by age band), race/ethnicity , education , BMI ; women in the lowest quartile of zinc intake, iron intake, or servings of grains; women who consumed < 1.5 servings of vegetables/day; and women in the OS. Tests for the statistical significance of interactions, though, are not reported.To improve comparability with previous cohorts that have evaluated dietary cadmium and cancer incidence, we applied the same methods for each outcome of interest to selected subgroups of women: never-users of hormone therapy (at enrollment); women with BMI between 18.5 and 25 kg/mn = 5,161 cases), women were censored at incidence of ER–, borderline, or unclassified breast cancer; conversely, for analyses specific to ER– breast cancer (n = 948 cases), women were censored at incidence of ER+, borderline, or unclassified breast cancer.In further analyses, we estimated associations between dietary cadmium and breast cancer cases classified according to estrogen receptor status (ER+ or ER–). For analysis restricted to ER+ breast cancer .Estimated dietary cadmium ranged from 0.02 to 59.4 μg/day , and was higher among women reporting higher levels of vegetable and grain consumption, consistent with foods documented to be high in cadmium, or higher energy intake . On averR2 of 0.4–0.5 for each in univariate analysis). Addition of zinc and iron intake (milligrams per day) to the model did not substantially change results; nor did substitution of daily total fiber (grams) and carbohydrates (grams) for vegetable and grain servings (data not shown). Not adjusting for total dietary energy intake assessed from the FFQ left the interpretation of results substantially unchanged, although some HRs for individual quintiles of dietary cadmium were significantly > 1 for endometrial and ovarian cancer in some but not all models . Adjustment for pack-years of smoking did not materially change results (data not shown).Results adjusted only for total energy, age, and WHI study component did not suggest statistically significant dose–response trends in associations between dietary cadmium and any of the three cancers . Further2; women without diabetes at enrollment; women consuming < 2.5 medium servings of fruits and vegetables and/or < 2.9 medium servings of grains per day (the lowest quartiles); women with zinc intake < 9.0 mg/day (the lowest quartile) from both diet and supplements; women with iron intake < 10.5 mg/day (the lowest quartile) from both diet and supplements; never-users of hormone therapy before enrollment; never-smokers; women with no history of any cancer before enrollment; and participants in the OS only (data not shown). Associations of ER+ and ER– breast cancer subtypes with dietary cadmium intake were similar to overall results for breast cancer (data not shown).Associations among the following subgroups of women were generally consistent with those estimated for the cohort as a whole: women with BMI 18.5–25 kg/mp = 0.007 for test of null hypothesis that ρ = 0).Mean creatinine-corrected urinary cadmium was 0.49 μg cadmium/g creatinine. The Spearman rank partial correlation coefficient (ρ) between energy-adjusted dietary cadmium and creatinine-corrected urinary cadmium, adjusted for age, was 0.085 . Although urinary cadmium has been used to measure low-level environmental exposure in many epidemiological studies, recent reports have suggested that urinary cadmium may not accurately reflect cadmium accumulation in the kidney resulting from long-term, low-level exposure; therefore, urinary cadmium may reflect primarily recent exposure rather than cumulative exposure , 2013. TFinally, although we did not assess occupational exposure to cadmium in the WHI, a previous study of the U.S. adult population suggests that elevated cadmium exposure occurs mainly in automotive and electrical repair, mining, metalworking, and similar jobs working directly with metals . BecauseDespite these limitations, our study has substantial strengths, including its prospective design and the large size of the cohort. Our study included triple the number of cases of breast and endometrial cancer risk and nearly double the number of ovarian cancer cases of the largest prior studies from the Swedish Mammography Cohort , 2012. FThe results of our study did not support the hypothesis that cadmium contamination of food, measured with an assessment of usual diet during the 3 months before baseline, is a risk factor for postmenopausal breast, endometrial, or ovarian cancer, but misclassification of exposure may have attenuated an association. In future prospective studies, alternative assessments of cadmium exposure, such as urinary cadmium concentration, should be tested in relation to risk of hormonal cancers.Program Office : Jacques Rossouw, Shari Ludlam, Dale Burwen, Joan McGowan, Leslie Ford, and Nancy Geller. Clinical Coordinating Center : Garnet Anderson, Ross Prentice, Andrea LaCroix, and Charles Kooperberg. Investigators and Academic Centers: JoAnn E. Manson , Barbara V. Howard , Marcia L. Stefanick , Rebecca Jackson , Cynthia A. Thomson , Jean Wactawski-Wende , Marian Limacher , Robert Wallace , Lewis Kuller , and Sally Shumaker .(139 KB) PDFClick here for additional data file.


Introduction. High BMI is a risk factor for upper body breast cancer-related lymphedema (BCRL) onset. Black cancer survivors are more likely to have high BMI than White cancer survivors. While observational analyses suggest up to 2.2 times increased risk of BCRL onset for Black breast cancer survivors, no studies have explored race or other social factors that may affect BCRL severity, operationalized by interlimb volume difference (ILD). Materials and Methods. ILD was measured by perometry for 296 overweight (25 > BMI < 50) Black (n = 102) or White (n = 194) breast cancer survivors (>6 months from treatment) in the WISER Survivor trial. Multivariable linear regression examined associations between social and physical factors and ILD. Results. Neither Black race  nor BMI  was associated with ILD. Attending college  was the strongest factor associated with ILD, followed by having more lymph nodes removed , >25% BCRL care adherence , and years since treatment . Discussion. Neither race nor BMI was associated with ILD among overweight cancer survivors. Education, a proxy for resource level, was the strongest factor associated with greater ILD. Tailoring physical activity and weight loss interventions designed to address BCRL severity by resource rather than race should be considered. Obesity is a major risk factor for onset of upper body breast cancer-related lymphedema (BCRL) –3, whichRace is a social construct , 24 and Identifying a profile for risk of greater BCRL severity is recognized as a great need in research , 15, 36,NCT01515124 at clinicaltrials.gov for details). Participants were recruited from southeastern PA and NJ zip codes, first through letters sent to patients diagnosed with breast cancer from 1999 to 2014 listed in each state's cancer registry or regional hospital-based cancer registry records and then through community events for breast cancer survivors.The study population is drawn from 351 women enrolled in the ongoing Women in Steady Exercise Research (WISER) Survivor trial conducted within the University of Pennsylvania's Transdisciplinary Research on Energetics and Cancer (TREC) Center. The WISER Survivor trial was approved by the University of Pennsylvania Institutional Review Board and all participants provided informed consent. WISER Survivor is a one-year randomized controlled weight loss and exercise intervention trial in sedentary overweight breast cancer survivors with breast cancer-related lymphedema  having stable (no cellulitis or flare-ups within the past 3 months) breast cancer-related lymphedema according to a standardized clinical evaluation by a Certified Lymphedema Therapist  and inteInterlimb arm volume difference was measured by perometry, operationalized as the difference between the left and right arms . The OptThe demographic, physical, and clinical data were collected via self-report survey at study entry and included race, age, education level (nonoverlapping categories), employment status, total number of persons living in the household, presence of dependent children at home, years since breast cancer diagnosis, type of breast cancer treatment(s), and history of BCRL treatment. Education level cut-offs were based on previous examinations of education, self-rated health, and obesity, which use four categories . However, due to sample size, the categories of less than high school and high school graduate were combined, resulting in a 3-category education variable . BMI was R: A Language and Environment for Statistical Computing. R Foundation for Statistical Computing, Vienna, Austria. URL https://www.R-project.org/). Missing data on cancer stage of diagnosis and bilateral BCRL were imputed via iterative regression. Multivariable linear regression models were constructed with an outcome of percent change in continuous interlimb volume difference, as a measure of BCRL severity. Models including a covariate for hand dominance [p < 0.05.Data analyses for this cross-sectional observational study were conducted using R version 3.2.2 (R Core Team (2015).ominance  were expp = 0.37); however, Black women had significantly higher BMI than White women .p = 0.89) nor BMI  was associated with interlimb volume difference. Completing some time in college was associated with a 4.9% smaller interlimb volume difference  though being a college graduate was not significantly different compared to completing high school or less (p = 0.28). Each year that passed since completing cancer treatment was associated with a 0.55% higher interlimb difference (p < 0.001), and receipt of axillary lymph node dissection was associated with an increased 4.75% interlimb difference when compared to sentinel lymph node biopsy (p = 0.01). Receipt of or adherence to lymphedema care treatment  was significantly associated with greater interlimb difference. Models (not presented) including a covariate for arm dominance did not change results.In This analysis of breast cancer-related lymphedema (BCRL) in overweight Black and White women found no differences in interlimb arm volume by race or BMI but found large differences by education, suggesting a greater role for resource than race for BCRL severity. Interlimb volume difference is a measure of BCRL severity used by previous researchers and as part of clinical criteria , 20. TheThe lack of difference in interlimb arm volume by race may mirror findings from a growing number of studies showing that the effects of race on BCRL onset are adjusted away in multivariable analysis  and thatPrevious explorations of BCRL development have focused on the same clinical and physical risk factors that predispose certain populations to greater BCRL onset, such as node dissection type and BMI , but havIn the United States, educational attainment is considered to be a proxy measure for the larger construct of socioeconomic position, with low attainment being highly correlated with low income, high poverty risk , and redLow-resource women may not have what they need to navigate worsening BCRL, especially in light of other competing demands that they face in day-to-day life. Still, in line with other studies with similar populations, adherence to BCRL care did not eliminate the interlimb volume disparity , 46. ThiContrary to some studies that have shown that BCRL can dissipate on its own in the first 18–24 months after treatment , our anaBehavioral interventions to reduce arm swelling have focused on the use of physical activity and weight loss to stabilize or lower BCRL severity , 5, 22. While this study boasts a large community-based sample with broad coverage across a metropolitan area, results may not be generalizable, due to the use of a sample of participants who were healthy enough to enroll in a clinical trial. The study sample may be healthier or better-resourced than the overall population of cancer survivors. Nearly half of the study sample had a college education, compared to 29.2% of women over 35 in the US (US Census 2014 estimates), meaning that these women may be more resourced than their peers and that work and household indicators may not represent stressors for the relatively high-resource women who comprised the study population. Since it is not a population-based sample, determining whether or not differences in BCRL severity exist by race or other social factors in the general population is not possible; however, no existing population-based datasets of the general population include information on BCRL severity, making this study the best available sample to assess these relationships. Although perometry is a reliable tool for assessing upper body interlimb volume differences, the use of perometry for measuring interlimb knee  and handThis dataset was limited to a few proxy measures of socioeconomic position and resource level and did not have data on income or direct measures of psychosocial stress. Education is an imperfect proxy of resource level or socioeconomic position, which may explain the nonmonotonic relationship between education and BCRL severity that has been observed in other analyses of education, obesity, and health outcomes for population-based adult samples . NeverthThis study was designed to explore differences by race in interlimb arm volume differences among breast cancer survivors with breast cancer-related lymphedema (BCRL), accounting for social and physical risk factors for BCRL. In this sample of overweight women, neither race nor BMI was associated with greater interlimb volume, but the social factor of education level had the strongest relationship with interlimb differences. These results suggest a greater role of resource than race for BCRL severity and point to a possible approach to reduce disparities in the burden of BCRL. Interventions focused on reducing interlimb volume may be differentially useful across resource levels. This may stand in contrast to interventions focused on BCRL onset, which may still warrant specific focus on Black women. Future studies should further examine the roles of education level and socioeconomic differences with expanded and more precise measures of education and socioeconomic position.


Elevated mammographic density (MD) is a strong breast cancer risk factor but the mechanisms underlying the association are poorly understood. High MD and breast cancer risk may reflect cumulative exposures to factors that promote epithelial cell division. One marker of cellular replicative history is telomere length, but its association with MD is unknown. We investigated the relation of telomere length, a marker of cellular replicative history, with MD and biopsy diagnosis.2) was measured using thresholding software. Associations between log-transformed LTL and continuous MD measurements (volume and area) were evaluated using linear regression models adjusted for age and body mass index. Analyses were stratified by biopsy diagnosis: proliferative  or non-proliferative (benign or other non-proliferative benign diagnoses).One hundred and ninety-five women, ages 40–65, were clinically referred for image-guided breast biopsies at an academic facility in Vermont. Relative peripheral blood leukocyte telomere length (LTL) was measured using quantitative polymerase chain reaction. MD volume was quantified in cranio-caudal views of the breast contralateral to the primary diagnosis in digital mammograms using a breast density phantom, while MD area (cmn = 141) was 1.6 (SD = 0.9) vs. 1.2 (SD = 0.6) in those with non-proliferative diagnoses (n = 54) (P = 0.002). Mean percent MD volume did not differ by diagnosis (P = 0.69). LTL was not associated with MD in women with proliferative (P = 0.89) or non-proliferative (P = 0.48) diagnoses. However, LTL was associated with a significant increased risk of proliferative diagnosis .Mean relative LTL in women with proliferative disease (Our analysis of LTL did not find an association with MD. However, our findings suggest that LTL may be a marker of risk for proliferative pathology among women referred for biopsy based on breast imaging. Mammographic density (MD) is a radiological reflection of the fibroglandular content of the breast, which histologically corresponds to both increased epithelium and stroma . EpidemiBoth environmental and biologic factors are thought to be responsible for the variations in breast tissue composition that are reflected in inter-individual differences in the extent of MD . FactorsTelomeres are nucleoprotein structures composed of repetitive DNA sequences (TTAGGG) and the shelterin protein complex. They cap the ends of chromosomes and help maintain genetic stability. TTAGGG repeats are lost during cell division, shortening telomeric DNA. When the telomeric DNA reaches a critical length, cells may undergo senescence, apoptosis or, if tumor suppressive mechanisms are abrogated, neoplastic transformation. Shorter telomeres have been observed in breast epithelial tumor cells compared with adjacent non-malignant tissue , with thAs with MD, it has been suggested that shortening of telomeres could be a consequence of exposures that drive cell proliferation , 30. We The National Cancer Institute (NCI) Breast Radiology Evaluation and Study of Tissues (BREAST) Stamp Project is a cross-sectional molecular epidemiologic study of mammographic density undertaken at the University of Vermont College of Medicine and its affiliated academic medical center, Fletcher Allen Health Care (FAHC). The study design and methodology have been described previously . BrieflyThe study was approved by the NCI Special Studies Institutional Review Board (IRB) and the University of Vermont IRB. Participants provided written informed consent to be part of the study and completed a standard health history questionnaireA research coordinator administered a telephone interview to collect additional health information. On the day of the breast biopsy, a research coordinator measured participants’ height and weight, and participants were asked to donate a blood sample. The informed consent included providing access to medical records and mammographic images and to breast pathology specimens not needed for clinical care. Compensation of $50 was provided to participants who opted to donate blood (processed and frozen as serum and blood clot) and/or mouthwash samples .non-proliferative  or proliferative, including both atypical and neoplastic entities . Information about biopsy type and laterality was abstracted from pathology reports.Breast biopsy and surgical pathology reports were reviewed for all study participants. For the purposes of this analysis, diagnoses were classified as Mammograms were acquired on one of six full field digital mammography systems at FAHC. Raw images were encrypted and transferred to the University of California at San Francisco for quantitative volume and area density assessment. This analysis was restricted to pre-biopsy cranio-caudal views of the contralateral breast. For women who underwent bilateral breast biopsies, the breast contralateral to the primary pathologic diagnosis was selected for analysis. If more than one mammogram was available, then the mammogram taken closest in time prior to the breast biopsy date was selected.3) and percent fibroglandular tissue volume using Single X-ray Absorptiometry (SXA), as described previously [2) by setting a pixel threshold for dense tissue on the images. Percentage mammographic density was calculated by dividing the absolute dense breast area by the total breast area and multiplying by 100. For both area and volume density measures, distributions of density measures were examined and images with extreme values were reviewed visually for validation.Breast density was quantified as an absolute fibroglandular tissue volume , 900 nM reverse telomere primer (GGCTGGCCTTACCCTTACCCTTACCCTTACCCTTACCCT), and 0.8 U of Platinum Taq polymerase. The reaction proceeded for one cycle at 95 °C for 5 min, followed by 35 cycles at 95 °C for 15 s, and 54 °C for 30 s. The β-globin reaction mixture consisted of 1× PCR buffer, 2.5 mM MgCl2, 100,000 fold dilution of SyberGreen, 200 nM dNTP mix, 2 % DMSO, 300 nM forward β-globin primer (CACATGGCAAGAAGGTGCTGA), 700 nM reverse β-globin primer (ACAGTGCAGTTCACTCAG CTG), and 0.5 U of Platinum Taq polymerase. The β-globin reaction proceeded for one cycle at 95 °C for 5 min, followed by 35 cycles at 95 °C for 30 s, 58 °C for 30 s, and 72 °C for 45 s. Each 96-well plate contained a no template negative control and two separate 5-point standard curves ranging from 0.024 to 15 ng using leukocyte DNA. These standard curves allowed the PCR efficiency to be determined for each experimental run. Each of the 10 plates also included three samples isolated from a series of cell lines with known telomere lengths, ranging from 3 to 15 Kb, as determined by terminal restriction fragment analysis. Inclusion of these samples provided an additional quality control check. The coefficient of variation (CV) for this cell line series ranged from 1.3 to 6.8 % across plates. Samples were re-run if the CV of either the telomere or the β-globin reaction was equal or greater than 5 % or either the telomere or the β-globin values fell outside the range of the standard curve. The maximum CVs were 4.8 and 3.9 % for the β-globin and telomere reactions, respectively. The average β-globin threshold (Ct) value and the telomere Ct value were calculated from the β-globin and the telomere triplicate reactions, respectively. For each sample, the telomere of the experimental sample to the single copy gene (T/S) ratio (−dCt) was calculated by subtracting the β-globin Ct value from the telomere Ct value. The relative T/S ratio (−ddCt) was determined by subtracting the -dCt of the middle samples of the cell lines series from the -dCt of each unknown sample. The relative T/S ratios  were used in the analysis.Whole blood samples were collected using standard techniques, allowed to clot for 30 min, and processed at the FAHC General Clinical Research Center. Samples were centrifuged at 3000 rpm for 15 min, and the serum and clot fractions were frozen at −80 °C until shipment to SeraCare Life Sciences , where they were stored in liquid nitrogen. Leukocyte DNA was isolated from blood clots at SeraCare using phenol chloroform extraction methods and quantified at the Cancer Genomics Research Laboratory  with the QuantiFluor® dsDNA System (Promega) according to the manufacturer’s instructions. DNA in 500 ng aliquots was sent to Johns Hopkins University School of Medicine, where quantitative polymerase chain reaction (qPCR) was used to estimate the ratio of telomeric DNA to that of a single copy gene (β-globin) as previously described , with thWe restricted the study population to participants who had SXA volumetric MD measurements, donated blood and whose breast biopsies contained terminal duct lobular units (TDLUs) suitable for assessment of telomere lengths . Of the 465 participants who consented, 12 were not subsequently biopsied and were excluded; 338 (75 %) women donated blood and had clots with a volume ≥1.0 mL, of whom 212 also had breast tissues available for telomere length assessment. Twelve women were missing SXA density and were excluded. Characteristics of the remaining eligible 200 women as compared to and the rest of the participants in the BREAST Stamp Project were similar with the exception of BMI, which was lower in the women included in our analysis (data not shown). Of the 200 women whose DNA underwent qPCR for relative LTL assessment, two samples failed quality control on two separate runs and were excluded. In addition, three participants had a relative LTL that was larger than three standard deviations from the study population mean and were also excluded, resulting in a final analytic population of 195 women.χ2 test for categorical variables, except when values in cells where less than or equal to 5 in which case the Fisher’s exact test was used. The Spearman correlation coefficient was computed to examine the correlation between relative LTL with age stratified by pathological diagnosis. Logistic regression was used to compute the association between relative LTL (continuous) and risk of proliferative disease adjusting for age (continuous) and BMI (continuous), which are known to be strongly associated with relative LTL.Statistical differences in participant characteristics by biopsy diagnosis (proliferative vs. non-proliferative disease) were computed using the Wilcoxon rank-sum test for continuous measures and the P-values were two-sided and P ≤ 0.05 was considered statistically significant. All analyses were performed using the R software environment (version 3.0.2).Relative LTL was transformed using the natural logarithm to improve normality. Multivariate linear regression was used to estimate the relationship between log-transformed relative LTL and participant characteristics adjusting for age (continuous) and BMI (continuous). Relative LTL was then back-transformed to the original scale and geometric means are presented. Similarly, multivariate linear regression was computed to examine the relationship between MD and relative LTL, adjusting for the potential confounders of age (continuous) and BMI (continuous). In sensitivity analyses, we additionally adjusted for age at first birth  and menopausal hormone therapy use . Quantitative volume and area density measures were transformed by taking the square root to approximate a normal distribution. Relative LTL was modeled as either binary, by categorizing relative LTL at the median relative LTL levels in subjects with non-proliferative diagnoses, or as a log-transformed continuous variable. Adjusted means of mammographic density measures were back-transformed by squaring the results. n = 141) and the remainder had a non-proliferative biopsy diagnosis (n = 54) (Table The majority of participants were diagnosed with proliferative disease n = 41 and thP = 0.005) and were more likely to have used menopausal hormone therapy (P = 0.039). MD measures did not differ between women with proliferative versus non-proliferative diagnoses. On average, relative LTL was similar among women with different categories of proliferative diagnoses . However, women with proliferative disease had longer (unadjusted) mean relative LTL compared with those with non-proliferative diagnoses (Table P = 0.002). After adjustment for age and BMI, relative LTL was associated with a significant increased risk of proliferative disease (Odds Ratio (OR) = 2.46 per one unit increase of relative LTL, 95 % CI: 1.47, 4.42; Table Compared with women with non-proliferative diagnoses, women with proliferative diagnoses tended to be older (P = 0.03). We did not find relative LTL to be associated with other risk factors such as age at menarche, parity, age at first birth, menopausal status, age at menopause or first degree family history of breast cancer. There was a suggested association between longer relative LTL and use of postmenopausal hormone therapy (P = 0.048).Relative LTL was weakly and inversely correlated with age in women with proliferative and non-proliferative diagnosis Fig. . After aLinear regression models evaluating the relation between MD and relative LTL were adjusted for age and BMI, which were inversely associated with percent volumetric and area MD measures in the entire study population as well as in women with proliferative and non-proliferative diagnoses (data not shown). MD was not associated with relative LTL in our study population; this was true irrespective of the measure of MD  or biopsy diagnosis Table . SimilarIn this population of women ages 40–65 undergoing diagnostic image-guided biopsy, relative LTL was not associated with mammographic density. However, we observed a novel association between longer relative LTL and proliferative biopsy diagnoses.Different measures of MD have advantages and limitations ; therefoThe few studies examining associations between relative LTL and breast cancer risk factors in women without breast cancer have reported shorter telomeres with older age and higher BMI , 13, 44;In this population, we found that women with proliferative disease had longer relative LTL than women with non-proliferative disease. Women with different proliferative diagnoses (hyperplasia vs. in situ cancer vs. cancer) had similar mean LTL. Our study involved participants representing a range of breast biopsy diagnoses, which makes comparisons with prior studies that included a healthy control group challenging. Healthy control groups may include subjects with proliferative but non-cancer diagnoses, which are related to breast cancer risk. Nevertheless, our findings are consistent with some epidemiologic studies that have found that longer LTL is associated with increased breast cancer risk –21, but A strength of our study was the use of quantitative, reliable density measures that have been validated with respect to breast cancer risk. Limitations include the relatively small sample size and narrower age range of participants than in most prior studies of LTL . It is pThis is the first study to examine the relationship between MD and relative LTL, and importantly, in a population that may be at higher risk of breast cancer, for which biomarkers of risk may be critical. While relative LTL was not associated with MD, we found that longer relative LTL was associated with proliferative lesions among women undergoing diagnostic image-guided biopsy. This finding suggests that LTL may be a marker of risk for proliferative pathology among women with abnormal breast imaging prompting a biopsy. Additional studies with larger populations and broader age ranges are warranted.


Single-nucleotide polymorphisms (SNPs) at 6q25.1 that are associated with breast cancer susceptibility have been identified in several genome-wide association studies (GWASs). However, the exact causal variants in this region have not been clarified.CCDC170 and ESR1 gene regions at 6q25.1 and accessed their associations with risk of breast cancer in a study of 1,064 cases and 1,073 cancer-free controls in Chinese women. The biological function of the risk variant was further evaluated by performing laboratory experiments.In the present study, we genotyped six potentially functional single-nucleotide polymorphisms (SNPs) within the CCDC170 and rs2228480 and rs3798758 in ESR1—with variant allele attributed odds ratios (ORs) of 1.38 : 1.20 to 1.57, P = 2.21 × 10-6), 0.84  and 1.19 , respectively. The functional variant rs9383935 is in high linkage disequilibrium (LD) with GWAS-reported top-hit SNP (rs2046210), but only rs9383935 showed a strong independent effect in conditional regression analysis. The rs9383935 risk allele A showed decreased activity of reporter gene in both the MCF-7 and BT-474 breast cancer cell lines, which might be due to an altered binding capacity of miR-27a to the 3' untranslated region (3' UTR) sequence of CCDC170. Real-time quantitative reverse transcription PCR confirmed the correlation between rs9383935 genotypes and CCDC170 expression levels.Breast cancer risk was significantly associated with three SNPs located at 6q25.1—rs9383935 in CCDC170, may be one candidate of the causal variants at 6q25.1 that modulate the risk of breast cancer.The results of this study suggest that the functional variant rs9383935, located at the 3' UTR of The online version of this article (doi:10.1186/s13058-014-0422-x) contains supplementary material, which is available to authorized users. Breast cancer is the most common female malignancy worldwide, accounting for 23% of all the new cancer cases in 2008 . The incCCDC170, also called C6orf97) and estrogen receptor 1 (ESR1) at 6q25.1, was first reported to be associated with the risk of breast cancer in Chinese populations . Ne. NeESR1 ogenesis . HoweverThis study included 1,064 breast cancer cases and 1,073 cancer-free controls, whose characteristics were described previously . Brieflyr2 > 0.8), which were further functionally evaluated by SNPinfo . Fo. Foet alll lines . miR-27all lines -28] and andet alll lines . Furtherll lines .CCDC170 expression. On the basis of the above results, rs9383935 could be considered a potential causal variant at 6q25.1. Additional studies are needed to validate our findings and extend the role of CCDC170 in the etiology of breast cancers.Taken together, these findings suggest that the variant rs9383935 may modulate individual susceptibility to breast cancer, possibly through regulating miR-27a inhibition of ESR1 and breast cancer risk . As. AsESR1 cer risk . Clinicacer risk . In a prtibility . In a fos1801132 .ESR1 in a larger sample size comprising 1,064 cases and 1,073 controls. We found significant associations of rs3798758 and rs2228480 with breast cancer risk Chinese women. The SNP rs2228480, located in exon 8 of ESR1, is a synonymous variant. The exon 8 involves in the assembling of the C-terminal region of ERα [et al. showed that diverse mRNA secondary structures with different stability were correlated to protein expression levels [As an extension of our previous studies, in the present study we evaluated potentially functional SNPs of n of ERα . This ren of ERα . Althougn of ERα . For exan levels . Anothern levels . NeverthCCDC170 rs9383935 showed the most prominent effect of any other variants of ESR1 or rs2046210, which provides new evidence for the role of the 6q25.1 region in breast cancer susceptibility. Further functional investigations of CCDC170 and 6q25.1 are warranted to fully reveal the mechanisms underlying the observed association with risk of breast cancer.Overall, in the present study, we evaluated six potentially functional SNPs within the 6q25.1 region and confirmed that rs9383935, rs3798758 and rs2228480 were associated with breast cancer in Chinese women, and we also replicated rs2046210 in accordance with previous reports. Specifically, the HS, XG and ZH conceived of the study and managed the overall project. YW, YH, ZQ, YJ and JC were responsible for sample-processing and performed the experiments. YW and GJ interpreted the initial statistical analyses and drafted the manuscript. JD and YW performed data management and checked the statistical analyses. HS, ZH, XG, GJ and HM contributed to the editing of the manuscripts. All authors read and approved the final manuscript.RMND1, C6orf211, CCDC170 and ESR1. The lower panel shows linkage disequilibrium (LD) plots of 31 SNPs in LD with rs2046210 and 6 selected functional SNPs marked with an asterisk. LD values between SNPs as indicated in the diamonds were measured by r2 in Chinese descent (CHB). For example, the r2 value between rs2046210 and rs9383935 was 0.86 in CHB. (JPEG 665 KB)Additional file 1: Figure S1.: Overview of the 6q25.1 region (chr6:151767683-152466099) from the UCSC Genome Browser (NCBI36/hg18). The upper panel shows the 698.4-kb region in 6q25.1 contain four genes: CCDC170 3' UTR luciferase reporter plasmid. (JPEG 219 KB)Additional file 2: Figure S2.: The construction of CCDC170 3' UTR. The figures and the values of minimum free energy (MFE) were generated in RNAhybrid (http://bibiserv.techfak.uni-bielefeld.de/rnahybrid/). Different alleles of rs9383935 are marked with squares. (JPEG 102 KB)Additional file 3: Figure S3.: The predicted binding affinity of miR-27a-3p and Authors’ original file for figure 1Authors’ original file for figure 2Authors’ original file for figure 3Below are the links to the authors’ original submitted files for images.


In most resource constrained settings like Nigeria, breast self-examination self-breast examination (BSE) is culturally acceptable, religious friendly and attracts no cost. Women's knowledge and beliefs about breast cancer and its management may contribute significantly to medical help-seeking behaviours. This study aimed to assess knowledge and beliefs of BSE among market women.A descriptive cross-sectional study was conducted among 603 market women in Ibadan, Nigeria. Data was collected using semi-structured interviews and analyzed using descriptive and analytic statistical methods.The mean age of the respondents was 34.6±9.3 years with 40% of the women aged between 30-39years. The proportion of married women was 339 (68.5%) with 425 (70.8%) respondents reporting that they do not know how to perform BSE. However, 372 (61.7%) women strongly agreed that BSE is a method of screening for breast cancer. Highest proportion 219 (36.3%) reported that the best time for a woman to perform BSE was ‘anytime’. Most of the respondents believed breast cancer is a dangerous disease that kills fast and requires a lot of money for treatment.More efforts are needed in creating awareness and advocacy campaigns in the grassroots in order to detect early breast cancer and enhance prevention strategies that would reduce the burden of breast cancer in Nigeria. Breast cancer is a global health concern and a leading cause of morbidity and mortality among all the cancers that affect women . In 2008The high incidence of breast cancer necessitates the need for early detection because this would increase the treatment options available to affected women and thereby improve survival rates . Some stScreening for early detection and diagnosis of diseases and health conditions is an important public health principle . Breast There are reports suggesting that factors related to women's knowledge and beliefs about breast cancer and its management may contribute significantly to medical help-seeking behaviours –22. ReceEthical approval was given by the Oyo state, Ministry of Health Ethics Committee in August, 2012. Participant information was anonymized and de-identified prior to analysis. Informed consent was obtained from the participants aged 15 and above. The named ethics committees approved the consent procedure in addition to the study protocol.A cross-sectional study was conducted between July to October 2012, in order to assess knowledge and beliefs on BSE among women selling in a few major markets in Ibadan, Oyo State, Nigeria. Major markets were purposefully selected and subsequently consented participants were interviewed from each of the market. Ibadan is the largest indigenous city south of the Sahara and is the capital of Oyo state, Nigeria. It has a population of about 2.6 million people . The stup) of knowledge of breast self-examination among market women = 0.5Estimate of the expected proportion (d) = 0.05Desired level of absolute precision (DEFF) = 1.5Estimated design effect  version 20. Descriptive and Chi-square statistics was employed in analysing the data.A total of 603 market women were recruited. More than three-quarters of the participants responded to knowledge on how to perform BSE.The majority of participants 425 (70.8%) reported that they did not know how to perform BSE, while only 29.2% reported that they do.Knowledge about how to perform BSE was slightly higher in participants who came from Oja-oba- market (37.6%) followed by those from Agbeni, Bode and other markets with level of knowledge all above 25% except in participants from Oje market. Very few participants were recruited in the ages below 20, and 50 years or above. Therefore the latter will not be considered further in this discussion, only ages 20–49 years will be reported. The percentage of participants who reported that they knew how to perform BSE increased with age in participants up to 49 years; with age group 40–49 years (40.4%) being the highest, followed by 30–39 years (32.0%) then 20–29 years. Married participants had a higher knowledge of how to perform BSE than single participants. Knowledge of BSE was correlated with educational level , and theIn total 271 participants responded to the question of when is the right time to perform BSE. Only 8.1% of these knew correctly that ‘mid-cycle’ was the right time to perform BSE.The highest proportion 219 (80.8%), reported incorrectly that the right time for a woman to perform BSE was ‘anytime’. Although below 10%, a large number of women who knew when to perform BSE came from Oja-oba market (8.3%) compared to the other three markets. However, these differences were not statistically significant. Knowledge about when to perform BSE decreased with increasing age showing a slightly higher level of knowledge in the 20–29 year olds than in the 30–39 and 40–49 year age groups(p<0.00). Compared to married, a double percentage of single women knew when to perform BSE. The level of knowledge about when to perform BSE was higher among post-secondary education (15.1%) while among other groups it was less than 10%. These differences were not statistically significant .About two-third (61.7%) of the study population strongly agreed that BSE is a screening method for breast cancer. About 28.5% agreed that fear of detecting breast cancer would make them not practice BSE, while more than 50% strongly disagreed with this statement. The majority of the women strongly disagreed that SBE should be done ‘only if you feel abnormal around your breast’. There was similar responses of participants who strongly agreed or disagreed about postures for SBE .When askThe present confirm findings from studies conducted in Nigeria over the past years, on BSE and breast cancer among women in the south-east , south , 32, souImportant knowledge deficits can have a detrimental impact on the education of women on screening practices and attitudes of women in the adoption of early detection practices . A correAssessment of the participants’ knowledge, attitude and beliefs showed that majority of the respondents reported that the right time to perform self-breast examination was ‘anytime’ and majority also disagreed that BSE should be done only when they feel abnormal around the breast. Some of the respondents reported that they would not practice BSE because they are afraid of detecting any evidence suggestive of breast cancer. In addition, findings from the study reveal that most of the study population have heard of breast cancer as a disease and self-breast examination as a screening method but there is still inadequate knowledge and understanding of the disease and its screening method. These findings are similar to the study among market women in Abakaliki (south-east Nigeria) by , women iThere is a great need for more awareness campaigns, advocacy to improve the knowledge of self-breast examination directed towards women of low socio-economic status and people at the grassroots levels in the country. This will ensure early detection and intervention to prevent mortality due to breast cancer.We conclude that knowledge about how and the time to perform BSE among Nigerian women working in the markets in Ibadan, south west Nigeria is very poor, particularly among women who are single, young with a low level of education after controlling for confounders. Similarly, few participants had strong negative beliefs towards breast cancer. Any interventions aimed at improving the knowledge about BSE and breast cancer screening should target these groups. Such interventions should be evaluated to ensure their success in improving women’s health.The quantitative nature of our study limits the extent to which the information reflects the nuanced views of respondents. A qualitative interview would have allowed a deeper understanding of the perspectives of the respondents.S1 Questionnaire(PDF)Click here for additional data file.