The mean serum level of leptin was 33.22 ± 21.35 ng/mL in patients and 29.49 ± 23.27 ng/mL in the normal participants (p = 0.3). Although, despite the magnitude of the associations, the results suggested no statistically significant contribution of -2548 G/A polymorphism (in leptin gene), Q223R polymorphism (in leptin receptor gene), and serum leptin levels in predicting the risk of breast cancer, further studies with larger sample size are suggested.Recent studies have shown that polymorphisms in leptin and leptin receptor genes are associated with increased risk for breast cancer. This study aimed at investigating -2548 G/A polymorphism in leptin gene and Q223R polymorphism in leptin receptor gene in patients with breast cancer. The study included 45 women with breast cancer and 41 healthy women. PCR-RFLP was used to determine the genotype of the subjects in terms of -2548 G/A polymorphism in leptin gene and Q223R polymorphism in leptin receptor gene. Serum levels of leptin were also measured by ELISA. For -2548 G/A polymorphism, the genotypes were homozygous AA (OR = 1.13;  The pression , 14. Thepression . Leptin pression  and mouspression , and CD2 systems . Long ch systems .Leptin is essential for the growth of human mammary gland and recent studies have shown that it may be effective in breast cancer . Leptin Various polymorphisms occur in leptin and its receptor gene including -2548 G/A and Q223R. In -2548 G/A polymorphism, an adenine replaces guanine in nucleotides -2548 upstream from the ATG start site at 5′ region of leptin gene promoter. This polymorphism affects the expression of leptin and its secretion from adipose tissue . Q223R pForty-five women with breast cancer and 41 healthy women were enrolled in this study. The patients included those women who were diagnosed according to pathology and the signs of the disease, while the controls were those with negative pathology results for breast cancer. All subjects were informed about the use of the results and voluntary participation in the study, and written consents were obtained from the patients and controls. 5 mL blood sample was collected for performing PCR-RFLP and measuring leptin. Clinical information about patients, such as cancer stage, estrogen and progesterone receptors status, HER2 status, and age, was recorded.μL of whole human blood treated with EDTA was used for extracting of genomic DNA. All centrifugation steps are carried out at room temperature (15–25°C). 20 μL QIAGEN Protease (or proteinase K) was transferred into the bottom of a 1.5 mL microtube and 200 μL whole blood was added to the tube and then 200 μL lysis buffer (AL buffer of the kit) was added to the sample and mixed by pulse-vortexing for 15 s and incubated at 56°C for 10 min. Then 200 μL ethanol (96–100%) was added to the sample and mixed again by pulse-vortexing for 15 s. Carefully the mixture was applied to the QIAamp Mini spin column (in a 2 mL collection tube) and centrifuged at 6000 ×g (8000 rpm) for 1 min. The QIAamp Mini spin column was placed in a clean 2 mL collection tube and the tube containing the filtrate was discarded. Carefully the QIAamp Mini spin column was opened and 500 μL Buffer AW1 (included in the kit) was added and then centrifuged at 6000 ×g (8000 rpm) for 1 min. Again the QIAamp Mini spin column was placed in a clean 2 mL collection tube and the collection tube containing the filtrate was discarded, and 500 μL Buffer AW2 (included in the kit) was added and centrifuged at full speed  for 3 min. The QIAamp Mini spin column was placed in a clean 1.5 mL microtube and the collection tube containing the filtrate was discarded and 200 μL Buffer AE (included in the kit) added and then incubated at room temperature for 1 min and centrifuged at 6000 ×g (8000 rpm) for 1 min. A second elution step with a further 200 μL Buffer AE increased yields by up to 15%. The average yield of DNA was 5.5 μg and A260/A280 ratio was 1.68–1.92.DNA was extracted using the QIAamp DNA Mini Kit . Briefly according to the procedure of this kit, 200 PCR-RFLP was performed for determining the genotype. Leptin concentration was measured by Mediagnost ELISA kit (Germany).μL reaction volume were as follows: 2.5 μL PCR buffer, 1 μL MgCl2, 0.5 μL dNTP mix, 1 μL of each primer, 3 μL template DNA, and 0.3 μL Taq DNA Polymerase .The primers indicated in  MspI enzyme (New England Biolabs) for leptin receptor gene and Hha1 enzyme for leptin gene for 16 h. After the incubation period, the products were electrophoresed again on 2% agarose. Regarding Q223R LEPR polymorphism, the presence of only one band of 416 bp length indicated individuals with QQ homozygous genotype, the presence of three bands of 416 bp, 291 bp, and 125 bp length indicated individuals with QR heterozygous genotype, and the presence of two bands of 125 bp and 291 bp length indicated individuals with RR homozygous genotype.PCR program for both polymorphisms was as follows: initial denaturation at 94°C for 4 minutes; 40 cycles including denaturation at 94°C for 30 s, annealing at 58°C for 30 s, and extension at 72°C for 30 s; and final extension at 72°C for 10 min. To confirm amplification of the desired fragments, all samples were electrophoresed on 1.5% agarose gel after PCR. The samples were then digested withRegarding -2548 G/A LEP polymorphism, the presence of only one band of 241 bp length indicated individuals with GG homozygous genotype, the presence of three bands of 241 bp, 181 bp, and 60 bp length indicated individuals with GA heterozygous genotype, and the presence of two bands of 181 bp and 60 bp length indicated individuals with AA homozygous genotype.t-test was used to compare the means and ANOVA and logistic regression were used to study the relationship between different genotype of -2548 G/A LEP polymorphism and Q223R LEPR polymorphism with breast cancer and leptin level. In all calculations, the probability level of p < 0.05 was deemed statistically significant. Statistical analyses were performed using SPSS-16.This study consisted of 45 women with breast cancer (mean age = 47.09 ± 11.45 years) and 41 normal subjects (mean age = 48.37 ± 8.80 years).p = 0.3).The mean level of serum leptin was 33.22 ± 21.35 ng/mL in the patient group and 29.49 ± 23.27 ng/mL in the normal group. Comparison of serum leptin levels in patients and controls revealed no significant difference in the serum level of this hormone between the healthy subjects and the patients . No significant association was found between the risk of breast cancer and homozygous GG, homozygous AA , and heterozygous GA .According to p = 0.33). No significant association was found between homozygous QQ with homozygous RR  and with heterozygous QR  and risk of breast cancer.Regarding genotype distribution of Q223R LEPR polymorphism in normal subjects, 6 were homozygous RR (14.6%), 18 heterozygous QR (43.9%), and 17 homozygous QQ (41.5%). The genotype distribution of Q223R LEPR polymorphism in patients was 1 homozygous RR (2.2%), 25 heterozygous QR (55.6%), and 19 homozygous QQ (42.2%). According to p > 0.05).The relationship between different genotypes of -2548 G/A LEP and Q223R LEPR polymorphisms with plasma leptin levels was investigated  and no sp > 0.05).In addition, the frequency of -2548 G/A LEP and Q223R LEPR polymorphisms and pathologic parameters in breast cancer patients  was not The present study examined the relationship between -2548 G/A polymorphism in leptin gene and Q223R polymorphism in leptin receptor gene with the risk for breast cancer in Yasuj, Iran.Breast cancer is the most common type of malignancy and the second leading cause of cancer death in women, so that its mortality rate was estimated approximately 25% until April 2013 . AdipokiThe effect of these polymorphisms on the risk of breast cancer has been studied in different populations. Woo et al. 2006) found no significant correlation between four polymorphisms in leptin receptor gene and the risk of breast cancer. They also observed no significant difference in serum leptin levels between patients and controls. The results of this study showed that the difference of mean serum leptin levels between various genotype groups of Q223R polymorphism was not significant. In the present study also no significant correlation was found between Q223R polymorphism and breast cancer risk; in addition, no significant difference was observed in the mean serum leptin levels between patients and controls and the mean difference in serum leptin levels was not significant between different genotype groups of Q223R polymorphism . Consist found noThe results of our study showed that -2548 G/A LEP and Q223R LEPR polymorphisms were highly polymorphic in the study population. The frequency of allele R in Q223R LEPR polymorphism was 0.366 in our study population which was higher than the figures reported in Pima Indian (0.32) and Arab (0.34) populations and lower than the figures reported in Caucasian (0.45) and Asian (0.85) populations .However, Chung et al. reported a little difference in R allele frequency of Q223R LEPR polymorphism according to racial groups . It is bThe differences in the findings of similar studies and the present study can be attributed to various genetic polymorphisms in similar metabolic pathways (gene-gene interaction). Other factors such as environmental factors  can affect leptin expression as confounders  and alteIn this study, no significant association was found between -2548 G/A LEP polymorphism, Q223R LEPR polymorphism, and serum leptin concentrations with the risk of breast cancer in the study population. Further studies are required to accept or reject the conclusion and to achieve more comprehensive results. In this regard, similar studies with larger sample size are recommended.


I2 = 96.7%, P < 0.001) compared with participants with one or more live births. Nonlinear dose-response association was found between parity and all-cause mortality (P for non-linearity < 0.0001). Our findings suggest that moderate-level parity is inversely associated with all-cause mortality.To quantitatively assess the association between parity and all-cause mortality, we conducted a meta-analysis of cohort studies. Relevant reports were identified from PubMed and Embase databases. Cohort studies with relative risks (RRs) and 95% confidence intervals (CIs) of all-cause mortality in three or more categories of parity were eligible. Eighteen articles with 2,813,418 participants were included. Results showed that participants with no live birth had higher risk of all-cause mortality (RR= 1.19, 95% CI = 1.03–1.38;  Multiple studies have shown that reproduction factors may affect the health of women in later life12345Drosophila melanogaster, described the existence of a trade-off between reproductive and somatic maintenance, suggesting that parity is associated with increased risk of death101112241“Antagonistic pleiotropy” and “disposable soma” theories, derived from experiments on both male and female Results of the literature search and selection are shown in 123456915161718192021222324I2 = 96.7%, P < 0.001) compared with that with 1 or more live births  for null parity, 1.15  for 1 live birth, 0.99  for 3 live births, 1.04  for 4 live births, and 1.12  for 5 or more live births .Eleven articles were included in our dose-response meta-analysis. Statistically significant evidence of non-linear association was found between parity and all-cause mortality . For men, people with high parity possibly have lower socioeconomic status, and risk of death is consequently higher because of poor access to healthcare servicesOur dose-response meta-analysis suggested that a J-shaped nonlinear association exists between parity and all-cause mortality in both women and men, which is consistent with previous studies145The observed statistical heterogeneity among studies could be attributed mainly to the diversity of study locations. Studies on Norwegians, Israelites, and Japanese showed large statistical heterogeneity. The larger proportions of higher-order planned births in more recent Norwegian cohorts because of higher availability of contraception, legal abortion, and “family friendly” policies than in earlier Norwegian studies may be a plausible reason for the large statistical heterogeneity observed among Norwegian studiesOur study has several strengths. First, most studies included in the analysis were cohort studies of large sample sizes and with long follow-up durations. This significantly minimized selection bias and considerably increased statistical power to detect potential association between parity and all-cause mortality. Second, subgroup and sensitivity analyses were used to investigate the source of statistical heterogeneity observed in our findings in great detail. Third, the previous meta-analysiset al. reported that a small proportion of step-parenthood was included in the population they studied, and that step-parenthood did not alter their findings on parity-related mortality among malesOur study has several limitations that should be addressed. First, non-marital birth or childbirth in previous marriage may be misclassified as null parity. This may have caused underestimation of the level of parity-related all-death risk. As for studies not reporting the precise number of births, we utilized categories of number of children for analysis in order to minimize misclassification. Additionally, biological parenthood and step-parenthood were not described in detail and were reported separately, which may also have reduced the risk estimate. However, Keizer In conclusion, results of our meta-analysis suggest that an association exists between parity and all-cause mortality. Low to moderate parity is associated with decreased risk of total death in both women and men. People with 3–4 live births have the lowest risk of total death. More prospective studies that control for all major confounding factors as well as studies exploring the biological mechanisms underlying the effect of parity on death risk are still needed.Studies published before 24 April 2015 were searched in the PubMed and Embase databases. The search was limited to studies carried out in humans, and the following key words and Medical Subject Headings were used: (“parity” or “number of live birth” or “number of children” or “parities” or “number of deliveries” or “number of living birth” or “number of live births” or “number of livebirth” or “number of kids” or “number of kid”) AND . To find additional references, we manually searched the bibliographies of all retrieved studies and selected all relevant publications. Only studies published in English were included. The following were excluded: conference literature, unpublished literature, and gray literature produced at all levels of government, academics, business, and industry in print and electronic formats, but not controlled by commercial publishersPublished studies were included in the meta-analysis if they met the following criteria: 1) the study was either prospective or had a historical cohort study design; 2) parity was the subject of interest; 3) the outcome was all-cause mortality; 4) the investigators reported relative risk (RR), hazard ratio (HR), or odds ratio (OR) and the corresponding 95% confidence intervals (CIs) for each parity category. If multiple publications had the same population as subjects, we included the most recent and most complete study. Two independent investigators (YZ and ZN) conducted initial screening of all titles or abstracts and then evaluated all potentially relevant articles based on full-text reviews.Two investigators (SL and XG) independently performed eligibility evaluation, data extraction, and quality assessment of each eligible study. All disagreements were discussed and resolved by consensus. The following data were extracted from each study: surname of the first author, publication year, study location, sex, and age range of the studied population, duration (in years) of follow-up, number of deaths, size of cohort, parity assessment, outcome assessment, parity category, RR or HR or OR and the 95% CI for each parity category, and factors adjusted in the report. If multiple estimates of the association were available, we extracted the estimate and adjusted for most covariates. If no adjusted risk estimate was presented, we used the crude risk estimate. If no risk estimate was reported, we calculated the crude risk estimate and its 95% CI using raw data provided with the article.Quality assessment was conducted according to the Newcastle–Ottawa quality assessment scale33In our meta-analysis, we used pooled RRs and their 95% CIs to measure the association between parity and all-cause mortality. Any result stratified by age and sex was treated as a separate report.I2 statisticI2 values of 25%, 50%, and 75% were assigned to low, moderate, and large statistical heterogeneities, respectivelyI2was <50%, whereas the random effect model was used when I2was ≥50%Statistical heterogeneity among studies was evaluated using Cochran’s Q test and P value was calculated by testing whether the two regression coefficients were simultaneously equal to zero. A null hypothesis stating that coefficient of the second spline is equal to zero was created to test the non-linearity. Original parity levels reported in each study were used in the dose-response analysis. If parity intervals were provided, the midpoint between the lower and upper bounds of parity interval was regarded as the corresponding parity dose. In cases that involve an open-ended upper interval, we assumed that the category exhibited the same amplitude as the adjacent interval7For dose-response analysis, we used the two-stage random-effect dose-response meta-analysis method proposed by Greenland and Longnecker37383939Moreover, sensitivity analyses were conducted to examine the influence of specific studies or high parity categories on overall RRs by excluding specific studies or data points with parity level above five live births. Publication bias was evaluated by conducting Egger’s and Begg’s regression tests44How to cite this article: Zeng, Y. et al. Parity and All-cause Mortality in Women and Men: A Dose-Response Meta-Analysis of Cohort Studies. Sci. Rep. 6, 19351; doi: 10.1038/srep19351 (2016).


CC vs. TT=1.01, 95% CI=0.96-1.06, Ph=1.00; ORCC+CT vs. TT=1.00, 95% CI=0.96-1.05, Ph=1.00; ORCC vs. CT+TT=1.02, 95% CI=0.98-1.07, Ph=0.94; OR allele C vs. allele T=1.01, 95% CI=0.98-1.04, Ph=0.99; ORCT vs. TT=1.01, 95% CI=0.93-1.09, Ph=1.00). Such lack of association with breast cancer was also observed in subgroup analyses according to ethnicity as well as in the analysis by source of controls. In conclusion, this meta-analysis suggests that the functionally important +936 C/T polymorphism may not be associated with breast cancer risk. Larger well-designed studies with gene-to-gene and gene-to-environment interactions are clearly required to validate the results further.The vascular endothelial growth factor (VEGF) gene single-nucleotide polymorphism involved in the regulation of the protein levels has been implicated in breast cancer. However, the published studies have produced contentious and controversial results. Herein, we performed a meta-analysis (from January to October 2013); to further evaluate the association between +936 C/T polymorphism and the risk of breast cancer. By searching the EMBASE, PubMed, and Web of Science databases, we identified a total of 12 case-control studies with 8,979 cancer patients and 9,180 healthy controls. The strength of the association was assessed using Odds Ratios (ORs) with 95% Confidence Intervals (CI). We found no evidence indicating that the allelic model or the genotype models of +936 C/T polymorphism were associated with the risk of breast cancer in total population (OR There has been much speculation that inhibition of VEGF activity may prevent tumor initiation, progression, and migration.4Angiogenesis is critical for the growth and metastasis of invasive tumors and constitutes an important component in the suppression of cancer formation.5,6 and breast cancer.7,8 The highly polymorphic VEGF gene comprises eight exons that produce different proteins by alternative splicing of a unique transcript generated from a single-copy gene.9 There are more than 100 single nucleotide polymorphisms (SNPs) identified in the VEGF gene to date (http://www.ncbi.nlm.nih.gov/SNP). A couple of polymorphisms within the region have been reported to be associated with the development of cutaneous malignant melanoma,10 lung cancer,11 and prostate cancer.12Several lines of work have shown a significant involvement of VEGF in human carcinogenesis. Enhanced VEGF expression and increased intratumoral microvessel density are related to an advanced stage disease and worse prognosis for a variety of women-specific malignancies, such as ovarian cancer13 and it is this relation that makes the polymorphism an ideal target for cancer research. A recent report indicated that +936 C/T polymorphism plays a major role in a broad range of human cancers, including breast cancer,14 a polygenic malignancy accounting for almost 16% of all cancer cases and 22.9% of female-specific diseases (http://www.who.int/cancer/detection/breast cancer/en/index1.html). However, current data on the association between VEGF +936 C/T and breast cancer susceptibility have shown a great discrepancy.15-26 Most importantly, the overall evidence of several meta-analyses, in which the included data are under intensive debate, suggested the association of +936 C/T polymorphism and breast cancer risk is not statistically significant.27-31 These findings seem to contradict the experimental evidence supplied in previous research. In the present study, we aimed to assess the debatable association further by means of meta-analysis based on corrected data and recently published studies to provide convincing evidence for the genetic contribution of +936 C/T genotypes in the risk of developing breast cancer.The most extensively studied SNP has been a C-to-T transition in the 3’-untranslated region (+936 C/T). A connection between +936 C/T status and VEGF plasma levels has previously been established,This meta-analysis was performed according to Moose Guidelines from January to October 2013 in Jiangsu Province (Nantong University), China. Literature SearchWe carried out a systematic literature search for manuscripts reporting on +936 C/T polymorphism and the risk of breast cancer in EMBASE (http://www.embase.com/), PubMed (http://www.ncbi.nlm.nih.gov/pubmed), and Web of Science (http://newisiknowledge.com) without any language restriction, using a combination of the following search terms: “vascular endothelial growth factor” or “VEGF”, “+936 C/T” or “rs3025039”, “genotypes” or “variant” or “variants” or “polymorphism”or “polymorphisms”, and “breast cancer” (the last search updated in June 2013). We also hand-searched the reference lists cited in genetic association studies, narrative reviews and meta-analyses to further identify the relevant publications that may have missed in the electronic search. We additionally carried out hand searches of the journals known to publish SNP-breast cancer articles, including International Journal of Cancer, Medical Oncology, Breast Cancer Research, Cancer Letters, Clinical Cancer Research, Cancer Epidemiology, Biomarker & Prevention, Breast Cancer Research Treatment, et al. All titles, abstracts, and full-texts were carefully reviewed to check if there were usable data reported in the original articles. When the required data were absent, corresponding authors were contacted via e-mail. However, no response was received by this process. Inclusion and Exclusion CriteriaThe following inclusion criteria were defined for the eligible studies in this meta-analysis: (a) the study must be published as a full-text paper with a case-control or cohort design; (b) the authors must have investigated the association between +936 C/T polymorphism and the risk of breast cancer; (c) the study must provide sufficient genetic data that could help to calculate an odds ratio (OR) and its 95% confidence intervals (CI); (d) the publication must be released online before the literature search was completed. We excluded the studies if; only breast cancer patients were investigated, information on genotype count was not detailed, or no responses were received from major authors. We also did not consider editorials, case reports, systematic reviews, and comment letters. In case of studies containing overlapped subjects, we selected the largest study with accessible genotype data.Data ExtractionIn order to maximize the reliability of data analyzed in this work, the data extraction was done by two independent reviewers (J Li and Yy Ding) based on a consensus reached on all items. The following characteristics were collected from each eligible article: first author’s surname, journal and year of publication, country where the study was carried out, ethnicity or racial descent (categorized as Caucasian or Asian), source of controls , baseline characteristics of the breast cancer patients and the disease-free individuals , assays utilized to genotype the polymorphism of interest, total cases and controls, genotype frequencies between cases and controls, matching status, adjusted factors and study design (categorized as retrospective or prospective). Accuracy of the extracted information was examined by crosscheck. Discrepancies, if any, were resolved through discussion with a senior reviewer (Yj Ju). Quality Assessment32 We defined the studies whose genotype distribution in controls was consistent with HWE (P>0.10) as high-quality studies, and those inconsistent with HWE as (P≤0.10) low-quality studies. We then conducted subgroup analyses according to quality  to confirm whether deviation from HWE had an influence on the association being investigated.The quality of included studies was assessed independently by the same two reviewers (J Li and Yy Ding) according to the P values of Hardy-Weinberg Equilibrium (HWE) in control groups.Statistical MethodsFor the controls of each study, HWE was examined by the chi-square goodness-of-ﬁt test with P<0.05 representing a significant deviation. The OR and its 95% CI were calculated to assess the strength of the association between +936 C/T polymorphism and breast cancer risk. The pooled ORs were performed for allelic model: allele C vs. allele T, and genotype models: CC vs. TT (homozygote), CC+CT vs. TT (dominant), CC vs. CT+TT (recessive), and CT vs. TT (heterozygote). The Z test was applied to determine the statistical significance of summary ORs, and P<0.05 was deemed statistically significant. 2 index indicating the proportion of inter-study variance attributable to heterogeneity rather than chance was used to measure the degree of inconsistency across studies. I2 statistics range from 0 and 100%, with higher proportions representing larger heterogeneity: 0-25% low heterogeneity, 25%-50% moderate heterogeneity, 50%-75% large heterogeneity, 75%-100% extremely large heterogeneity. The ﬁxed effect model derived from the Mantel-Haenszel method33 was used to pool the ORs when P>0.10 or I2 statistics less than 50%; otherwise, the random effect model described by Darlington34 was more appropriate. In addition, subgroup analyses according to ethnicity and source of controls were performed to detect the potential heterogeneity further.Heterogeneity assumption was evaluated by a chi-square based Q test. The significant level was defined at P<0.10. In addition, the I35 An asymmetric plot and the P values of Egger’s linear regression test smaller than 0.10 suggested obvious publication bias.The influence of each individual study on the summary ORs was evaluated by excluding the single studies in turn from the pooled estimate. Publication bias was examined visually by funnel-inverted plots in which the standard error of log (OR) of each study was plotted against its log (OR). Egger’s test was utilized to check the asymmetry of funnel plots on the natural logarithm scale of the ORs.All statistical data were analyzed with Stata software . All tests were two-sided with a statistically significant level at P<0.1. Selection and Characteristics of Eligible Studies15-26 for this meta-analysis, involving 8,979 breast cancer cases and 9,180 cancer-free controls. Although we primarily obtained 697 records, the initial removal of obviously irrelevant records and duplicates resulted in 27 studies for further evaluation. We subsequently deleted 15 studies due to various reasons, including research on other SNPs at the same locus, lack of data for calculation of ORs, case-case study, and comment letters.36-50 The main characteristics of the included studies are summarized in As depicted in Meta-Analysis Resultsh=1.00). To test whether there is a positive association in another contrast models we subsequently assumed the CC+CT vs. TT model, failing to find any significant association . In terms of the remaining models tested, the statistical data showed no clear associations with the genetic risk of breast cancer   see .CC vs. TT=1.00, 95% CI=0.95-1.05, Ph=1.0; ORCC+CT vs. TT=1.00, 95% CI=0.95-1.05, Ph=1.0; ORCC vs. CT+TT=1.02, 95% CI=0.97-1.07, Ph=0.87; OR allele C vs. allele T=1.01, 95% CI=0.98-1.04, Ph=0.99; ORCT vs. TT=0.99, 95% CI=0.90-1.08, Ph=1.0, and Asian populations ORCC vs. TT=1.04, 95% CI=0.93-1.16, Ph=0.75; ORCC + CT vs. TT=1.02, 95% CI=0.93-1.12, Ph=0.79; ORCC vs. CT+TT=1.02, 95% CI=0.92-1.13, Ph=0.74; OR allele C vs. allele T=1.02, 95% CI=0.96-1.10, Ph=0.68; ORCT vs. TT=1.07, 95% CI=0.91-1.26, Ph=0.70. Similarly, the subgroups of hospital-based studies and population-based studies did not indicate any statistical evidence of significant breast cancer risk in relation to +936 C/T polymorphism at either the allelic or the genotypic level, nor did the final analysis according to the quality of study reveal a statistically significant association . In the Begg’s test, no evidence of publication bias was detected by visual inspection of the funnel plots. P values of Egger’s test also indicated that there was no significant publication bias across the included studies  see .51 The potentially functional polymorphism in the VEGF gene, +936 C/T, was shown to correlate with lower VEGF production.13,15 It is the biological function of +936 C/T that leads to widespread attention towards the association between presence of +936 C/T polymorphism and breast cancer susceptibility. The published reports, however, have produced contentious and controversial results.15-17 Heterogeneous populations across the earlier association studies are a plausible explanation for the controversy. Other factors, such as varying study designs, number of individuals’ genotypes, different experimental methodologies may also contribute to the existing discrepancy among the studies.Angiogenesis is a complicated process that could result in the formation of new blood vessels from pre-existing vasculature and thereby facilitates tumor growth, invasiveness, and development of metastasis. Data from experimental and clinical research have suggested that breast cancer is an angiogenesis-dependent disease closely associated with the serum level of VEGF.Meta-analysis is an analytical tool that provides credible evidence for malignancy risk and well-characterized sequential variations by systematically summarizing existing data. In the current meta-analysis, we combined all available data on the association of +936 C/T polymorphism with breast cancer risk. The pooled analysis showed that there was no increased or decreased risk of breast cancer in correlation with any genotype or allele at +936 C/T. The lack of association with breast cancer was also observed in subgroup analyses according to ethnicity. In subsequent stratification analyses by source of controls and quality of study, nor did we find a clear association, the same as the previous analyses. 27 who analyzed 5,729 cases and 5,868 controls, found +936 C/T polymorphism may not contribute to breast cancer susceptibility. This study was later commented and the association was re-evaluated in a recent publication, in which several methodological issues were addressed.28 The re-evaluation nevertheless failed to find any significant association. Consistent with the two earlier assessments, several meta-analyses reported in 2011 did not lend support modified breast cancer risk in relation to +936 C/T genotypes.29-31Several meta-analyses have been carried out, in an attempt to identify the association of +936 C/T polymorphism with breast cancer risk. For instance, Gu et al.,One shared issue among these meta-analyses is that not all of the published data are included. In addition, there was substantial heterogeneity in each of the earlier meta-analyses, leading to an increased likelihood of biased results and hence making overall estimates less reliable. We enlarged our sample by incorporating all available data of the precious meta-analyses and the recently published studies, which helped to strengthen the statistical power and increased the precision of our findings, although we did not find any evidence of significant correlation. Another explanation may be that the polymorphism being investigated is a low-penetrance genetic variant and the minor impact on breast cancer susceptibility cannot be detected unless a sufficiently large analysis is performed. It is therefore worthwhile to determine the genetic association in a large-scale study. 52,53 It has been shown that the +936T allele associated with reduced levels of VEGF abrogates activator protein 4 located in +936C allele, and thereby results in reduced VEGF expression.13 The +936T allele was also shown to have major effects on decreased uptake of 18F-fluorodeoxyglucose, a marker used to detect and diagnose breast cancer.54. It is therefore convincing that VEGF +936 C/T likely play a significant role in breast cancer susceptibility. If this speculation can be confirmed in the following studies, the +936 C/T polymorphism may be used as a biomarker for breast cancer.According to experimental evidence of functional genetic polymorphism at candidate locus, the VEGF +936T allele is speculated to have an association with predisposition to breast cancer. Activator protein 4 represents a helix-loop-helix transcription factor that upregulates serum levels of various cellular and viral genes through binding to particular sites, and it is these specific enhancer sites that modulate expression of VEGF.We did not find a significant association between VEGF +936 C/T genotypes and breast cancer susceptibility in both Asian and Caucasian populations. The real association may be underestimated or even masked as a result of the small number of individuals in each ethnic group, the Asian group in particular . The minor allele frequency (+936 C/T) of Asian populations (18.2%) is higher as compared to the global frequency . Available data show a connection between the +936T allele and lower VEGF protein levels that is related to increased susceptibility of breast cancer, suggesting VEGF +936 C/T polymorphism may have effect modification of breast cancer in Asians. Our findings are obviously inconsistent with this hypothesis of a positive association and the inconsistency hints that the relation merits further investigation.23,26 and this may have influenced our results, despite no indicated apparent alterations in the data with or without them. Secondly, we carried out an exhaustive literature search in medicine-specific databases and supplemental hand-search, failing to identify the publications written in other languages except English. As a result, only published data in the English language were analyzed in this meta-analysis. Further, unpublished data and ongoing studies were not considered in the present work, which may lead to biased results. Third, crude OR instead of adjusted OR was calculated to assess the risk of breast cancer. Since there was no uniform standard definition for the independent studies and the calculation of ORs varies substantially across the published studies. For example, Luo et al. adjusted ORs for sex and age, whereas the adjustment factors in the study by Oliveira et al. were age and race. The wide difference makes it impossible to provide an OR based on adjusted factors. Finally, it is known that breast cancer is an etiologically heterogeneous malignancy. Genetic contribution alone only accounts a part for the breast pathogenesis and the exact molecular mechanism that underlies this invasive disease remain largely unknown. Available data documented a significant involvement of gene-to-gene and gene-to-environment interactions in the development and progression of breast cancer, which, however, cannot be confirmed in this analysis due to lack of the original data.Several limitations of this study should be considered in interpreting our results. Firstly, mild deviation from HWE that probably results from methodological issues such as inappropriate or erroneous genotyping, selection bias and population categorization was detected in two studies,The aforementioned shortcomings, nevertheless, could not overshadow the strong points of this quantitative assessment. We performed a meta-analysis with a maximum sample to date and it is the sample sufficiency that has contributed to robust and convincing findings. Besides, we updated earlier meta-analyses with new information from subsequent published studies and determined that +936 C/T genotypes were not involved in breast cancer incidence. Last but not least, publication bias and inter-study inconsistency (heterogeneity) are two major problems when performing meta-analysis. The confounding factors appeared to have no notable influence on this investigation, as suggested in the analytical methods.In summary, the accumulated evidence from prospective studies supports that +936 C/T polymorphism is not significantly associated with the risk of breast cancer. Well-designed studies with a larger number of samples and with gene-to-gene and gene-to-environment interactions are considered necessary to clarify the association. This meta-analysis suggests that the common +936 C/T polymorphism may not be associated with breast cancer risk. This finding merits further research where the sample size is substantially large, and gene-gene and gene-environment interactions are considered.


P = 0.034) as the frequency of C allele . We found similar results for men, but not for women . The results of this study show that the rs10046 (C/T) polymorphism of CYP19A1 gene exhibits correlation with CAD and that patients with C allele have an increased probability of manifesting the disease.The CYP19A1 gene encodes the enzyme aromatase, which is responsible for the biosynthesis of estrogens. The rs10046 polymorphism of CYP19A1 gene has been investigated in two studies on the occurrence of hypertension, but there are no studies on its correlation with coronary artery disease (CAD). We investigated 189 subjects who were hospitalized at “KAT” General Hospital of Athens and underwent coronary angiography. Of these, 123 were found with CAD with an average age of 60 years and constituted the patients group and 66 subjects with an average age of 58 years without damage in the coronary vessels and constituted the control group . The frequencies of genotypes CC, CT, and TT of rs10046 polymorphism are significantly different between the group of CAD patients and the control group (0.34, 0.48, and 0.18 versus 0.20, 0.48, and 0.32, resp.,  It is obvious that geneticists would mainly focus on investigating the correlation of genes with ischemic cardiovascular disease since it is the first cause of morbidity and mortality worldwide , 2. The The rs10046 polymorphism, a C/T single nucleotide polymorphism (SNP) located in the 3′ untranslated region (3′-UTR) of the CYP19A1 gene, has been related according to the number of T alleles either to trends of increased levels of circulating estrone and estradiol but no differences for androgens, either lower levels of several androgens but no differences for estrone or estradiol, or higher estrogen to androgen ratios . HoweverWe studied 189 subjects who were hospitalized for precordial pain investigation and underwent coronary angiography at the Cardiology Department of “KAT” General Hospital of Athens during 2012. All subjects were Greeks with no blood relationship. Of these, 123 were found with CAD with an average age of 60 years (81% men) and constituted the patients group. Coronary artery disease was defined as a luminal narrowing >70% of at least one vessel. The remaining 66 subjects with an average age of 58 years (78% men) exhibited no damage in the coronary vessels and constituted the control group . Subjects were defined hypertensive or diabetic if they were under treatment or their blood pressure was >140/90 mm Hg and fasting glucose >126 mg/dL, respectively. Smoking status was defined from current smokers. Ancient history of CAD was present in 19% of the patients' group and in 0% of the controls' group. Family history was considered positive for CAD patients if at least one first-degree relative was diagnosed with CAD by the age of 55 years. The data were obtained from medical files. All subjects provided written informed consent. The study was performed according to the Code of Ethics (Declaration of Helsinki).∧C sites (isoschizomers: SduI). The digestion of the PCR product by the restriction endonuclease Bsp1286 I (SduI) allowed discrimination between the alleles C and T of the CYP19A1 polymorphism. The 202 bp PCR product in the presence of C at 3′ UTR is cleaved into two fragments of 172 bp and 30 bp. The CC genotypes resulted in 172 bp and 30 bp, the CT genotypes in 202 bp, 172 bp, and 30 bp, and the TT genotypes in 202 bp digestion products  analysis. The PCR amplification primers used were forward, 5′-CTG GAA CAC TAG AGA AGG CTG GTC AGT GC-3′, and reverse, 5′-GTT CTC TGG TGT GAA CAG GAG CAG ATC AC-3′. The PCR procedure was as follows: an initial denaturation step at 94°C for 2 min and then amplified for 35 cycles at 94°C for 30 sec, 57°C for 40 sec, and 72°C for 40 sec, followed by a final extension step at 72°C for 7 min. The PCR products were digested with the restriction enzyme Bsp1286 I (SduI), were separated by electrophoresis on 1,5% (w/v) agarose gel, and were visualized by ethidium bromide staining. The digestion of the PCR product by the restriction endonuclease Bsp1286 I (SduI) allowed discrimination between the alleles C and T of the CYP19A1 polymorphism. The 202 bp PCR product in the presence of C at 3′ UTR is cleaved into two fragments of 172 bp and 30 bp. The CC genotypes resulted in 172 bp and 30 bp, the CT genotypes in 202 bp, 172 bp, and 30 bp, and the TT genotypes in 202 bp digestion products . The subn, %). Unpaired Student's t-test was used to evaluate differences in continuous variables once normality was demonstrated (Kolmogorov-Smirnov or Shapiro-Wilk test); otherwise, a nonparametric test (Mann-Whitney) was used. Differences for categorical variables were analyzed with chi-square test . Differences in genotype frequencies between CAD patients and controls were evaluated using the chi-square analysis for homogeneity. Differences in allele frequencies were tested for statistical significance at the 95% confidence interval using Fisher's exact test. A P value less than 0.05 was considered statistically significant. The odds ratio (OR) was used as a measure of the strength of the association between allele frequencies and CAD. Data analysis was performed with IBM SPSS 19 statistical software .Continuous variables are presented as mean ± SD and categorical variables as observed number . Similar findings were pointed for men with the same statistical significance for the genotypic frequencies (P = 0.034) and the C allele risk for CAD (OR = 1.881 and P = 0.010); females consisted of small group for patients and controls and did not demonstrate any statistical significance, although the C allele risk had a high odds ratio for the CAD patients  and a significant increased risk was found for patients with C allele (OR 1.771 and P = 0.0095), who have an increased probability of manifesting the disease. Similar findings were pointed for men, but not for women, although the C allele had a high odds ratio risk for the CAD patients. The latter result can be explained by the small size group for females. We may conclude that rs10046 polymorphism of CYP19A1 gene is a risk factor for the occurrence of CAD, but without knowing the exact mechanism. To this direction the measure of sex hormones levels would be undoubtedly useful. However, no other studies have been performed on this polymorphism and CAD and, taking into account the multifactors of the disease, it is necessary to significantly increase the sample and the research is continued and confirmed by genome-wide association studies (GWAS) in connection with aromatase activity.Our study was performed in two groups with similar risk factors, which were separated by coronary angiographic documentation in patients with CAD and healthy. The results of this study show that rs10046 (C/T) polymorphism of CYP19A1 gene exhibits correlation with CAD. The frequencies of CC, CT, and TT genotypes of rs10046 polymorphism demonstrated significant differences between the group of patients with CAD and the control group (Polymorphisms associated with the occurrence of CAD are risk factors for this disease and our efforts to prevent the onset or progress of the disease must be directed to the people bearing these polymorphisms in their genetic material.


Additional analyses were performed using the programs, multi-factor dimensionality reduction  and MDR permutation test (version 0.4.7), which consider all variables as risk factors simultaneously. The results of the present study demonstrate that the SLC23A2-05 and KRAS-LCS6 polymorphisms are not a risk factor for squamous cell carcinoma of the head and neck. In the same samples, the association of alcohol consumption (P<0.001) and smoking habit (P<0.001) with cancer presence was positive when each variable was considered individually. Concerning the environmental factors, a positive association of smoking habit and alcohol consumption with cancer, although not with ethnicity  was identified using the MDR tool, which analyzed the variables and polymorphism genotypes simultaneously. In conclusion, in the present study, squamous cell carcinoma of the head and neck was highly affected by environmental factors when compared with the affect of SLC23A2-05 and KRAS-LCS6 polymorphisms.Cancer is a genetic disease that is highly influenced by environmental factors. To determine the risk factors of squamous cell carcinoma of the head and neck, two polymorphisms, solute carrier family 23 member 2 (SLC23A2-05 [rs4987219]) and v-Ki-ras2 Kirsten rat sarcoma viral oncogene homolog (KRAS)-LCS6 (rs61764370), and environmental factors, including smoking and alcohol consumption, were studied in a population. The present study included 165 males diagnosed with squamous cell carcinoma of the head and neck. The control group consisted of 230 healthy male subjects without cancer or a family history of cancer. The SLC23A2-05 and KRAS-LCS6 polymorphisms were analyzed by polymerase chain reaction followed by enzymatic digestion. All patients and healthy subjects were assessed with regard to their smoking habit and alcohol consumption as these are considered to be risk factors for cancer. The statistical analysis was performed using logistic regression, Fisher’s exact and χ Cancer of the head and neck is associated with high morbidity and mortality rates of patients worldwide, including Brazil, and is a global health concern ,2. StudiKRAS2 gene  -LCS6 (G>T) polymorphism is located in the , KRASB)  and appe, KRASB) .RAS genes, HRAS, KRAS and NRAS, were reported in the 3′untranslated region (UTR) region with the expression regulated by the let-7 microRNA (miRNA). The let-7 miRNA expression was reduced in lung tumor tissue compared with normal tissue, however, the RAS protein expression was greater in lung tumors, which indicates a potential underlying mechanism of let-7 miRNA in cancer  variant possessed an increased risk of disease. Furthermore, the KRAS-LCS6 variant allele is associated with a high KRAS expression and decreased levels of let-7 -05 , which is localized in the SLC23A2 (nucleobase transporter) member 2, region 20p13-p12. The sodium-dependent SLC23A2 protein is responsible for the transport of vitamin C  and is cIn this context, the objective of the present study was to evaluate the prevalence of the KRAS-LCS6 and SLC23A2-05 polymorphisms in patients with squamous cell carcinoma of the head and neck, and to verify the significance of the polymorphisms on the involvement and severity of the disease via an association study.A cross-sectional study was performed according to a case-control model. The study was conducted by comparing the case  and control (subjects were paired by gender and age to the case subjects) groups. A negative family history of cancer was considered to be an inclusion criterion.There were 165 males diagnosed with squamous cell carcinoma of the head and neck that were treated at the Clinical Oncology and Otorhinolaryngology, Faculty of Medical Sciences, State University of Campinas . The DNA samples were stored at the Laboratory of Cancer Genetics . The control group comprised of 230 healthy male subjects.The present study was approved by the Ethics committee of the State University of Campinas  and written informed consent for participation in the study was obtained from the subjects.The extraction of DNA from peripheral blood leukocytes was performed according to standard protocol for the phenol-chloroform method , with spThe clinical variables analyzed in the samples were as follows: Cancer presence (control or case group), ethnicity (Caucasian or not), fatality, smoking habit , alcohol consumption , age at diagnosis, tumor-node-metastasis (TNM) staging system , degree of cancer differentiation  and localization. For the control group, the variables considered were smoker/alcoholic , ex-smoker/ex-alcoholic  and non-smoker/teetotal.Medical specialists assessed all of the variables, with special consideration given to TNM stage and the adenocarcinoma differentiation degree. The evaluated variables were normalized according to the literature .2 (25 mM), dNTP (25 mM for each nitrogenous base), primers (0.2 pmol of sense and antisense primers), Taq polymerase (5 units) and genomic DNA (50 ng/ml). The primers used were as follows: Sense, 5′-GGT GTC AGA GTC TCG CTC TT-3′ and antisense, 5′-GGG TCG TAT ACC AAA GGC CTT AG-3′ for the KRAS-LCS6 polymorphism (to amplify a fragment of 420 bp); sense, 5′-AAA TGC TCT GGG CAA CCT TA-3′ and antisense, 5′-CCC CCA GGA CAT CGA CAA-3′ for the SLC23A2-05 polymorphism (to amplify a fragment of 385 bp).The polymorphism analysis was by polymerase chain reaction (PCR) and enzymatic digestion. To determine the KRAS-LCS6 and SLC23A2-05 polymorphism genotypes the following PCR conditions were used: Bidistilled water, 10× Taq Buffer with KCl, MgClThe annealing temperature was 64.6 and 64.9°C for the KRAS-LCS6 and SLC23A2-05 polymorphisms, respectively, for one minute. The initial denaturation was at 95°C for 5 min, followed by 35 cycles of 94°C for 1 min  and 72°C for 2 min; the final amplification was at 72°C for 7 min.Hinf I enzyme  for the KRAS-LCS6 polymorphism and with Tai I  for the SLC23A2-05 polymorphism, at 37°C for 14 h according to the manufacturer’s instructions.Following PCR, the enzymatic digestion was performed with the The reaction was analyzed on a polyacrylamide gel (12%) with a voltage of 180 V for 4 h. The gel was stained in ethidium bromide solution and visualized on a Typhoon™ scanner .According to the fragments observed, the genotypes were identified as follows: i) KRAS-LCS6 polymorphism; TT (8+80+135+197 bp), TG (8+80+135+197+332 bp), and GG (8+80+332 bp) and ii) SLC23A2-05 polymorphism; CC (385 bp), CG (128+270 bp) and GG (128+270+385 bp).The mean age difference between the case and control groups was evaluated by Student’s t-test using the R program . The HarGenotypic and allelic association calculations were performed by logistic regression using the logistf function environment in the R program , conside2 test. Using the same program and tests, the association between the polymorphisms, and tumor localization and differentiation degree was assessed. A final analysis was performed concerning the combination of the two polymorphisms, KRAS-LCS6 and SLC23A2-05, and the presence of cancer.To associate cancer presence with smoking habit, alcohol consumption and ethnicity as risk factors, the Statistical Package for the Social Sciences version 17   was usedExtra data was obtained using the multi-factor dimensionality reduction (MDR) version 2.0 and MDR permutation test (PT) version 0.4.7 programs. The two programs were used in order to evaluate the genetic interaction between the polymorphisms in the samples. The MDR model is non-parametric and genetic model-free data mining for the identification of non-linear interactions among genetic and environmental attributes –26. To aThe statistical power of the samples was estimated using GPower 3.1.6 , to deteIn the case group  100 patients were diagnosed with laryngeal cancer, and 65 were diagnosed with throat (larynx and pharynx) cancer. With regard to smoking habit; 76.98, 21.21 and 1.81% of the subjects were smokers, ex-smokers and non-smokers, respectively, and for alcohol consumption; 55.76, 30.91 and 13.33% were alcoholics, ex-alcoholics or teetotal, respectively.In the control group , the smoking habit group comprised of 21.61, 41.71 and 36.68% smokers, ex-smokers and non-smokers, respectively, and for alcohol consumption, 9.55, 27.64 and 62.81% were alcoholics, ex-alcoholics or teetotal, respectively. Alcohol consumption was analyzed in 199 control subjects.In the KRAS-LCS6 polymorphism case group, 135 (81.82%) patients exhibited the TT genotype, 29 (17.58%) exhibited the TG genotype and 1 (0.61%) exhibited the GG genotype. In the control group, 187 (81.30%) exhibited the TT genotype, 41 (17.83%) exhibited the GT genotype and 2 (0.87%) exhibited the GG genotype. In the case group the allele frequency was 0.91 and 0.09, for the T and G allele, respectively. In the control group the allele frequency was 0.90 and 0.10, for the T and G allele, respectively.In the case group for the SLC23A2-05 polymorphism, 35 (21.21%) patients exhibited the CC genotype, 84 (50.91%) CG and 46 (27.88%) GG. In the control group, 43 (18.69%) exhibited the CC genotype, 132 (57.39%) CG and 55 (23.91%) GG. In the case and control groups the allele frequency was 0.47 and 0.53, for the C and G allele respectively.The KRAS-LCS6 polymorphism was in Hardy-Weinberg equilibrium in the case and control groups: P=0.9899 and P=0.9999, respectively; as were the SLC23A2-05 polymorphisms in the case and control groups; P=0.9232 and P=0.0332, respectively (data not shown).In The combination of the KRAS-LCS6 and SLC23A2-05 polymorphisms with cancer presence is shown in The MDR program analysis  did not Cancer is a complex disease and is a significant health problem. Cancer is a genetic disease, which is highly influenced by environmental factors. Determining the risk factors, including genes and environmental variables, improves current knowledge regarding cancer, provides improved tools for its diagnosis and treatment, as well as a method to determine the risk within a given population. In the present study, squamous cell carcinoma of the head and neck was analyzed considering two polymorphisms, SLC23A2-05 (rs4987219) and KRAS-LCS6 (rs61764370), and environmental factors .et al (et al (For the KRAS-LCS6 polymorphism, the case and control groups were in Hardy-Weinberg equilibrium. In the literature, there is a hypothesis that the activation of the KRAS pathway may occur in squamous cell carcinomas by the action of the KRAS-LCS6 normal variant, which may therefore be associated with the susceptibility and clinical outcome of cancer . The preet al  is not pl (et al  analyzedIn the present study the G allele occurred with a low frequency, in the case and control groups. Therefore, this allele is not considered to be a risk factor to disease susceptibility. The same was found when the sample was evaluated by cancer location. Furthermore, when correlated with the disease stage, no association was found, which indicates no influence on disease progression. However, in the present study the polymorphism was analyzed in the normal tissue of the case patients. Therefore, the hypothesis that this polymorphism may be significant in tumor development could not be excluded, and further investigation of this polymorphism in tumor tissue is required.In the present study, the SLC23A2-05 polymorphism in the case and control groups demonstrated the same allele frequency. The allele frequency of C and G were 0.47 and 0.53, respectively. In a previous study, C and G allele frequency in an African-American population was 0.80 and 0.20, and in Caucasians was 0.42 and 0.58, respectively . In a caet al (A study by Chen et al  evaluateIn TNM staging, there was no statistically significant association identified with the SLC32A2-05 polymorphism. In this context, these polymorphisms may not be cancer severity modulators. As for the KRAS-LCS6 polymorphism, the SLC32A2-05 polymorphism was analyzed in normal tissue.Concerning smoking habit and alcohol consumption, there was an association with cancer risk. In this context, the squamous cell carcinoma of the head and neck was influenced by environmental factors, however, there was not a modifier gene associated with the polymorphisms that were investigated in the present study.In conclusion, squamous cell carcinoma of the head and neck is a significant disease worldwide. Therefore, determining the factors that are associated with the severity and presence of this type of cancer is considered to be important. Globally, smoking habit and alcohol consumption are considered to be risk factors, resulting in individuals exhibiting squamous cell carcinoma of the head and neck. In the present study, the KRAS-LCS6 and SLC32A2-05 polymorphisms were not considered to be cancer risk factors, however, they were analyzed in the blood and not in tumor tissue. Potentially, the observation of these polymorphisms in a localized tumor may provide a significant tool for the detection of cancer. In future studies, the investigation of additional genes and tumor tissue is required to clarify the complex mechanisms associated with cancer, as well as protein quantification to determine additional data regarding cancer severity.


In view of previous conflicting findings, this meta-analysis was performed to comprehensively determine the overall strength of associations between brain-derived neurotrophic factor (BDNF) genetic polymorphism Val66Met and susceptibility to bipolar disorders (BPD).Literatures published and cited in Pubmed and Wanfang Data was searched with terms of ‘Val66Met’, ‘G196A’, ‘rs6265’, ‘BDNF’, ‘association’, and ‘bipolar disorder’ up to March 2014. All original case–control association studies were meta-analyzed with a pooled OR to estimate the risk and 95% confidence interval (CI) to reflect the magnitude of variance.Twenty-one case–control association studies met our criteria for the meta-analysis. Overall, there was no significant difference in allelic distribution of Val66Met polymorphism between patients and controls with a pooled OR = 1.03  although there was a trend towards association between Val66Met polymorphism and BPD in Caucasians with an OR of 1.08 . However, subgroup analyses showed that there was a significant association of Val allele with decreased disease susceptibility for bipolar disorder type II with a pooled OR of 0.88 .There is no compelling evidence to supportVal66Met polymorphism in BDNF gene playing an important role in the susceptibility to BPD across different ethnicities.The online version of this article (doi:10.1186/s12888-014-0366-9) contains supplementary material, which is available to authorized users. Bipolar disorders (BPD) are chronic, recurrent, debilitating disorders with high lifetime prevalence and significant disease burden across different populations -3. HowevThe hypothesis of neuronal plasticity involved in mood disorders has been supported by the use of antidepressants and mood stabilizers, e.g. lithium and valproate, inducing the expression of neurotrophins (e.g. BDNF) and synaptic changes ,17. Morehttp://www.ncbi.nlm.nih.gov/pubmed/) and Wanfang Data (http://www.wanfangdata.com/), with the keywords ‘Val66Met’, ‘G196A’, ‘rs6265’, ‘BDNF’, ‘association’, and ‘bipolar disorder’ in varying combinations. The retrieved abstracts were used to identify studies that examined the allelic association between the Val66Met polymorphism of BDNF and bipolar disorder. Bibliographies or citations from retrieved articles were also cross-referenced as well. The searched period was from the first data available in each database up to March 2014. Two independent authors extracted the following data from each eligible study: last name of the first author, year of publication, ethnicity, sample sizes and allele frequencies of cases and controls, etc. Discrepancies were resolved by mutual consent.Studies included in the analysis were searched from two databases: Pubmed (All eligible studies were determined against the following inclusion criteria: (i) published in a peer-reviewed journal; (ii) presented original data; (iii) provided either allele frequency of Val(G)/Met(A), or genotypes  in both BPD patients and healthy controls; (iv) enrolled more than 100 subjects in both patients group and controls group; and (v) designed as a case–control study. Both family-based studies and genome-wide association studies were excluded in this research. Duplications were deleted, as well as studies that reported all or part of their data previously. The authors of studies were contacted for additional information  if there was uncertainty about whether their data met our inclusion–exclusion criteria, or if we needed additional data which were not contained in the original report.Data were classified by diagnostic category (case or control) and allele , and Val was assigned as the risk allele. Meta-analysis was performed similar to that described previously . The pooThe process of identifying studied included in this meta-analysis is shown in Figure Overall, the data from 7219 BPD cases and 9832 healthy controls were analyzed. The mean genotype distribution in Caucasian and Oriental population was presented in Table Among 21 included studies, six case–control association studies made a distinction for clinical phenotypes between bipolar I disorder (BP I) and bipolar II disorder (BP II), and two other studies only recruited BP I patients Table . A furthWith a total of 7,219 patients and 9,832 control cases, our meta-analysis included an additional 4,076 BPD cases and 3,485 healthy controls compared to a previous meta-analysis of case–control studies . SimilarOur overall finding, that no convincing evidence for association between the Val66Met polymorphism and BPD as a whole, is consistent with a previous meta-analysis reported by Kanazawa et al. , which iThe finding of insignificant association between the Val66Met polymorphism and BPD in Oriental population populations , but a trend towards significance in Caucasian population suggest that ethnic heterogeneity may affect the results of these genetic association studies. Fan et al.  reportedThe finding of significant associations between Val66Met polymorphism and BP II suggests that the strength of association between Val66Met polymorphism and BPD could depend on the clinical phenotypes or subtypes of BPD such as with rapid cycling course, early onset, or substance comorbidity ,50,55,59Several limitations of this meta-analysis should be considered. One important limitation is that we only investigated relatively well-studied polymorphic variations in BDNF gene. Another limitation is that family-based studies and genome-wide association studies were excluded due to the heterogeneity of research methods. In a previous meta-analysis, the pooled OR derived from nine case–control studies is nominally significant , while the pooled OR derived from five family-based studies increases notably to an OR of 1.54 (P = 0.000019). Finally, since we primarily designed the current analysis to demonstrate the potential association(s) between Val66Met polymorphism and bipolar diagnostic boundaries, the association of this polymorphism with other characteristics of BPD such as sex, onset age, comorbidity, impairment in brain morphology and function, or treatment response, were not explored.Taken together, the Val66Met polymorphism in BDNF gene may be involved in the pathogenesis of BPD by influencing the susceptibility of specific subtypes such as BP II, but there is no compelling evidence of BDNF gene playing an important role in susceptibility to BPD across different ethnicities. The associations observed in current meta-analysis should be interpreted with caution. Further large-scale studies with same definitions of phenotypes and controls in homogenous ethnic groups are warranted to elucidate the relevance of BDNF gene variations as a risk factor for BPD (or diagnostic subtypes) susceptibility.


Furthermore, OR estimates in subgroup analyses were discussed, and strong associations were observed for dichlorodiphenyldichloroethylene , hexachlorocyclohexane , chlordane , and hexachlorobenzene . This meta-analysis had suggested that total OCPs of interest was significantly positively associated with NHL risk.Growing evidence indicates that exposure to organochlorine pesticides (OCPs) could increase non-Hodgkin lymphoma (NHL) risk. However, results from epidemiological studies investigating this association remain controversial. We thus conducted a meta-analysis to quantitatively evaluate the association between OCP exposure and NHL risk. Relevant publications were searched in PubMed, Web of Science, and Embase and identified according to the inclusion criteria. Thirteen studies  were selected for this meta-analysis. We used odds ratios (ORs) with 95% confidence intervals (CIs) to estimate the relationship between OCPs exposure and NHL risk. The summary OR for included studies was 1.40 (95% CI 1.27 to 1.56). No overall significant heterogeneity in the OR was observed (P It has been estimated that there will be 72,850 new cases  and 20,150 deaths  of NHL in the United States in 2016, accounting for 4.3% of all new cancer cases and 3.4% of all cancer deaths, respectivelyNon-Hodgkin lymphoma (NHL), a heterogeneous group of malignancies arising from the lymphatic system, including lymph nodes and lymphoid tissues and eventually migrating to the bone marrow and blood, is the sixth most commonly diagnosed cancer in both males and females in the United States1In fact, the incidence of NHL has nearly doubled in western countries since the 1970s, although no further increase has been found since the mid-1990s8Organochlorine pesticides (OCPs), one of the most abundant of the persistent organic pollutants (POPs), have been used extensively in agriculture as insecticides and for malaria prevention worldwide since the 1940s. Their properties, including persistence, long half-lives, high lipophilicity and long-range transport, and potential toxicity, have enabled them to be bioaccumulated in human tissue through the food chain and pose adverse impacts on human health111415et al.et al.719A number of studies have recently associated the risk of NHL with some environmental pollutants, including OCPs, with inconsistent results. For instance, Spinelli In the present study, we performed a meta-analysis to quantitatively evaluate whether the exposure to OCPs  was associated with the risk of NHL with large sample sizes on the basis of published epidemiological evidence.The results of literature research and selection progression were shown in h) of 0.253 and I2-value of 12.6%. As for associations of exposure to specific objects and the risk of NHL, 5 main classifications of OCPs, including DDT, DDE, HCH, chlordane, and HCB, were studied independently. According to our results, the exposure to DDE, HCH, chlordane, and HCB had been found to increase the risk of NHL , while non-significant association was confirmed between DDT exposure and NHL risk .Based on the results from the Cochrane Q-test with a P-value of 0.253 for heterogeneity, the fixed-effect model was applied to evaluate the risk of NHL related to OCPs. As shown in h = 0.024, I2 = 43.2%) and a low heterogeneity in nested case-control studies , but the association was not found between OCPs and NHL risk in case-cohort studies . When stratified by objects, DDE, HCH, chlordane, and HCB showed a significant association  with NHL risk, while a non-significant association was suggested for DDT . According to year of publication and geographical region, significantly increased associations between OCPs levels and NHL risk were found in subgroups of year of publication (≤2005 or >2005) and geographical region (Europe or Northern America). Within the subgroup analyses of sample type, the results showed that the level of OCPs exposure determined from the types of mixed (adipose tissue and blood) and adipose tissue alone had a higher and significant risk for NHL  than that from blood sample . In addition, the subgroup of quality score was examined, and there was significant association among the groups with moderate and high quality , but not with low quality .As shown in Meanwhile, to further evaluate the relationships among the exposures to the 5 main types of OCPs  and the risks for NHL, we conducted another subgroup analysis among these 5 objects by the group stratified by age, BMI, and number of adjustments, and detailed results had been shown in h = 0.253 and I2 = 12.6%. In particular, exclusion of the studies with non-adjustment, the largest sample size, the smallest sample size, deviation values, and low-quality scores was performed to assess reliability and stability based on fixed-effect and random-effect models, the results of which were consistent with the pooled OR of OCPs exposure levels on NHL risk. We also omitted 3, 4, and 5 studies in a random manner using both fixed-effect and random-effect models for further study. Leave-one-out analysis was conducted by excluding one study in turn, and the pooled ORs and 95% CIs did not change dramatically as shown in In order to evaluate the robustness of our conclusions about the exposure levels of OCPs on NHL risk, sensitivity analyses were conducted by different manners of exclusions as shown in The association of OCPs exposure and NHL risk was first suggested by a few epidemiological studies among farmers who were supposed to have a high possibility of exposure to OCPs during their agricultural work21232425262528h = 0.253, I2 = 12.6%). Besides, specific OCPs had been studied independently. The results had suggested that body burden levels of OCPs, including DDE, HCH, chlordane, and HCB, were significantly associated with increased risk of NHL , while DDT exposure showed no evidence for a significant link with NHL risk .In spite of the considerable effects from the epidemiological studies, the association between OCPs exposure and NHL risk has remained unclear. Thus, to evaluate the association between OCPs exposure and NHL risk, we conducted the present meta-analysis to pool the results from 13 relevant studies. As for the overall association, a significantly increased association between OCPs exposure and NHL risk was found , with a low heterogeneity of the included studies , while the nested case-control studies group had a larger risk than the case-control studies group, reaching a statistically significant level . The exposure data in nested case-control and case-cohort designs are obtained on a subset of the full cohort, which are the most common approaches for reducing the costs of exposure assessment in prospective epidemiological studies. Furthermore, the cohort design was seen as a more valid method than the case-control design, which was more prone to potential biases related to control selection and recall, thus cohort results are more reliableNHL was more common in developed areas, with the highest incidence rates found in Australia, Western and Northern Europe, and Northern America. The lowest NHL rates were found in Asia and Eastern Europe, except some special histological subtypes. Exactly, the geographical region of the studies in our meta-analysis mainly included Europe  and Northern America , with a distinct increased risk of NHL in Northern America  and Europe . However, we should note that studies from Asia and Africa were lacking in our meta-analysis, and these findings need to be interpreted with caution. Further comprehensive studies should be conducted in Asia and less developed countries to further determine the geographical difference in risk of NHL.2-value results from the blood subgroup than from adipose tissue, we believed that different definitions and methods of detection of total lipids might be potential reasons of heterogeneity, and the difference between serum and plasma might had affected the I2 test results in part. As for the group of mixed with adipose tissue and blood, it was not enough to examine our conclusions due to the limited studies.OCPs were reported to possess high lipophilicity, thus they could be bioaccumulated in human tissue through the food chain and pose a threat to human health11et al. between 1988 and 1995 showed a positive correlation between increased BMI and the risk of NHL, especially diffuse large cell lymphoma and follicular lymphomaet al. and raised the possibility that obesity was positively associated with an increased risk of NHL In addition to the difference among the group by sample type, high lipophilicity of OCPs was hypothesized to lead to an increased risk among individuals with higher BMI. A report in 2014 evaluated the association between pre-diagnostic obesity and survival for patients with NHL, with the conclusions implying that pre-diagnostic obesity might be related to poorer overall and NHL-specific survival in patients with NHLet al. observed a higher incidence rate in older age groups than in younger age groups for all birth cohorts examined, and recent cohorts had a higher incidence rate than earlier cohorts, especially among the older age groups in both men and women based on all NHL cases reported to the Connecticut Tumor Registry between 1935 and 1988Age was also suggested to be a major confounder in the association between the exposure to OCPs and the risk of NHL. For one side, OCPs were reported to possess long half-lives, enabling them to maintain a stable structure inside the human body and adversely affect human health for a long period. Under these circumstances, the exposure level of OCPs could increase with age, thus older individuals could have a higher level of OCPs exposure. For the other side, age was an essential factor in the occurrence and development of a variety of cancers, including NHL tumors. It had been suggested that age might play an essential role in the incidence of NHL, and the overall incidence of NHL increases steadily with age at diagnosisApart from the potential confounding effects of age and BMI, analysis on the numbers of adjustments was conducted. The total identified studies were stratified by the numbers of adjustments. Consequently, the results revealed that the groups with 3 or more adjustments  showed a lower risk than the groups with less than 3 adjustments . Results with more adjustments would be more likely to express the true association. Nonetheless, these findings confirmed the reliability of our results, which remained consistent even after the adjustments of major confounding factors.Several limitations and uncertainties of our meta-analysis should be acknowledged. Firstly, publication bias and selection bias might be inevitable due to the inclusion of only published studies and with the exclusion of non-English studies. Secondly, though most studies gave a sufficient exposure gradient in the NHL risk assessment, tertile, quartile, and quintile were used for the OR estimate via setting the lowest quantile as the reference. Therefore, the exposure measurement was different across the studies, which might be a potential source of heterogeneity in this analysis. Thirdly, the exposure to OCPs  with estrogenic properties might pose an increased NHL risk as a result of the interactions among these chemicals. Finally, since various subtypes of NHL had different characteristics and possibly diverse etiologies, the associations with overall NHL, rather than with subtypes of NHL, would lead to uncertain problems, thus the results should be interpreted with caution.In conclusion, the findings of our meta-analysis yielded a statistically significant association between exposure to OCPs and elevated risk of NHL. Subgroup analyses of DDE, chlordane, and HCB further confirmed the reliability of our results, which remained consistent even after adjustments of the major confounding factors (age and BMI). Future research with larger sample sizes is warranted to confirm these findings and to determine the likely biological mechanism.In this meta-analysis, the literatures were identified by a search on PubMed, Web of Science and Embase from the period 1986 to June 2015 for observational studies in relation to OCPs and NHL. Various combinations of the following keywords were used in our search strategy: “organochlorine”, “pesticide”, “DDT”, “DDE”, “HCB”, “HCH”, “chlordane” and “non-Hodgkin lymphoma”. Moreover, the reference lists of the selected articles were also reviewed for potentially eligible studies. The identified articles were limited to studies published in English. This meta-analysis had been conducted according to the PRISMA checklists and followed the guideline to report the resultsStudies were considered eligible in the meta-analysis if they satisfied the following criteria: (1) the study design was observational, (2) the exposure of interest was DDT, DDE, HCB, HCH, chlordane, (3) the measurement of the exposure of interest should be conducted in body fluids or tissues, not using questionnaires to review the exposure history, (4) the outcome of interest was NHL, (5) the study examined the association between DDT/DDE/HCB/HCH/chlordane exposure and NHL, and (6) the study should include reported ORs or relative risk (RR) with 95% CIs or provided sufficient information to extrapolate relevant values. Meanwhile, we excluded the studies that did not report original results, such as reviews, editorials, comments, correspondence, letters, meta-analyses, and meeting abstracts. We identified 797 studies through the systematic literature search, and finally 13 studies  were included in this meta-analysis.Data extraction was performed independently and reviewed by 2 reviewers (D.L. and T.Z.), and the results of this data extraction were compared between the 2 reviewers, and consensus was obtained before the meta-analysis. The reviewers extracted the following information for each study: first author name, publication year, country of study, type of study design, number of cases or controls, type of OCPs, sample type, gender, age of cases and controls, OR and corresponding 95% CI, and the confounding factors used for the adjustment of ORs. The ORs with more adjusted confounders were chosen when studies had different models for the calculation of estimated risks.According to the 9-star Newcastle–Ottawa Scale (NOS)Among the 13 identified studies, 4 studies741428171820444546472 statistic2 values of 25%, 50%, and 75% as low, moderate, and high degrees of heterogeneity, respectively, were used to assess heterogeneityHeterogeneity among the results of these studies was estimated using the Cochran Q test and ISubgroup analyses according to study design, object, year of publication, geographical region, sample type, quality score, and adjustments were conducted with the hope to evaluate the potential sources of heterogeneity and to investigate the detailed aspects of the association between OCPs levels and the risk of NHL.Sensitivity analyses were conducted to explore the effects of a specific study on the pooled OR. In addition to the Begg’s testAll statistical analyses were performed using the Stata version 12.0 software , other than quality-effect modeling . All statistical tests were two-tailed, and a P-value of <0.05 was considered statistically significant.How to cite this article: Luo, D. et al. Exposure to organochlorine pesticides and non-Hodgkin lymphoma: a meta-analysis of observational studies. Sci. Rep.6, 25768; doi: 10.1038/srep25768 (2016).


Mounting evidence from experimental and animal studies suggests that vitamin A may have a protective effect on melanoma, but the findings on the association of vitamin A intake with risk of melanoma have been inconsistently reported in epidemiologic studies. We attempted to elucidate the association by performing a meta-analysis.Eligible studies were identified by searching PubMed and EMBASE databases, as well as by reviewing the references of retrieved publications. Summary odds ratios (OR) with corresponding 95% confidence interval (CI) were computed with a random-effects model. Study-specific ORs and 95% CIs for the highest vs. lowest categories of vitamin A intake were pooled.A total of 8 case-control studies and 2 prospective studies comprising 3,328 melanoma cases and 233,295 non-case subjects were included. The summary OR for the highest compared with the lowest intake of total vitamin A, retinol and beta-carotene was 0.86 (95% CI = 0.59–1.25), 0.80 (95% CI = 0.69–0.92) and 0.87 (95%CI = 0.62–1.20), respectively. Significant heterogeneity was observed among studies on vitamin A and beta-carotene intake, but not among studies on retinol intake. Subgroup and sensitivity analyses confirmed these findings. There was no indication of publication bias.Findings from this meta-analysis suggest that intake of retinol, rather than of total vitamin A or beta-carotene, is significantly associated with reduced risk of melanoma. Malignant melanoma is a potentially fatal tumor that continues to increase in incidence, and the increase is considered real and not due to earlier detection or changes in diagnostic criteria The effect of vitamin A, an essential nutrient that cannot be synthesized by humans, on melanoma development is of particular interest. Retinoids (e.g. retinol) have been documented to be effective in inhibiting proliferation, inducing apoptosis and differentiation, as well as inhibiting growth of murine and human melanoma cell linesTo the best of our knowledge, there has been no comprehensive quantitative assessment of the association of vitamins A intake with risk of melanoma. In response, we took up this meta-analysis summarizing published case-control and cohort studies to further elucidate this issue.Potentially relevant publications were searched on PubMed and EMBASE databases through February, 2014 using the following search terms: (melanoma OR skin neoplasm OR skin cancer) AND (vitamin A OR retinol OR carotene OR provitamin A OR carotenoids) AND (cohort OR prospective OR retrospective OR nested OR case-control). In addition, the reference lists of retrieved publications were also carefully reviewed for any further studies.Studies were eligible for inclusion if they met the following inclusion criteria: (1) study design was case-control or cohort; (2) exposure of interest was vitamin A intake, or intake of two selected measures of vitamin A (retinol and beta-carotene); (3) outcome of interest was melanoma; and (4) odds ratio (OR), relative risk (RR) or hazard ratio (HR) with corresponding 95% confidence interval (CI) were reported. The titles and abstracts of all potentially relevant publications were reviewed to evaluate the relevance of the information; full texts were scrutinized if any potentially relevant information was identified in a retrieved abstract. For publications with same population resources or overlapping datasets, the largest or most recent one was included.Using a standardized data-collection protocol, the following data were extracted from each study: the first author's last name, publication year, country of origin, study design, age and sex, number of cases/non-cases, exposure details, ascertainment of exposure, risk estimates with corresponding 95% CIs for the highest compared with lowest levels of vitamin A intake, and variables accounted for in the analysis. If both the crude risk estimates and multivariate-adjusted ones were reported, the one reflecting the greatest degree of adjustment was extracted. Data abstraction was performed independently by two reviewers, and any discrepancy was discussed and adjudicated by a third reviewer until a consensus was achieved.A random-effects model taking into account both within- and between-study variation was assigned to compute the summary risk estimates I2 statistics P-value of less than. 1 was considered statistically significant heterogeneity. Potential publication bias was evaluated by using Egger's test and Begg's funnel plot Statistical heterogeneity was assessed with Q and The search strategy yielded 978 citations . Three aP = 0.02, I2 = 66.8%).P = 0.45, I2 = 0.0%) , when comparing those with the lowest intake. There was no heterogeneity among studies ( = 0.0%) .P<0.001, I2 = 71.9%) , with significant heterogeneity (= 71.9%) .Results of subgroup analysis are reported in P for Egger's test was 0.88, 0.31 and 0.69, respectively), which was supported by the visual inspection of the Begg's funnel plot.There was little indication of publication bias for studies on vitamin A, retinol or beta-carotene  reduction in the risk of melanoma, when comparing those with the lowest intake. These findings were confirmed by subgroup and sensitivity analyses.Our finding of a protective effect of retinol intake on melanoma risk was supported by evidence from experimental and animal studies. Retinol belongs to a class of compounds called retinoids Our finding of null association between beta-carotene intake and risk of melanoma was in line with results from a recently published meta-analysis of randomized controlled trials that reported no effect of beta-carotene supplementation on risk of melanoma  Individual studies are in general of low power to detect a significant association between exposure and outcome, especially when the incidence of disease was relatively low. The major strength of our study is the inclusion of large number of melanoma cases, which had largely enhanced the power of the study.Several limitations to this meta-analysis should also be noticed. First, most included studies are retrospective case-control studies. Exposure data in these studies were collected after case diagnosis, raising a problem of recall bias. In a study of diet and melanoma, however, it is unlikely that cases had differentially recalled their diet intakes, or that cases had changed their diet habits due to the diagnosis of the disease, because little knowledge of an association between diet and melanoma exists. However, selection bias is still a concern. Second, most included studies assessed diet intake with a self-reported FFQ, which may have resulted in measurement errors and contributed to exposure misclassification. Third, although included studies provided risk estimates accounting for multi-variables, including sun exposure related factors, some unmeasured or unknown confounding cannot completely be ruled out. Subjects with high intake of retinol may also be more likely to have other health-care behaviors that impacted melanoma risk. Fourth, dose-response analysis was not carried out because only few studies In summary, results from this meta-analysis indicate that intake of retinol, but not total vitamin A or beta-carotene, is significantly associated with reduced risk of melanoma. Our study solved the inconsistency of the literatures on the associations of vitamin A intake and risk of melanoma. However, the prevention of melanoma should still rely on reducing traditional risk factors for melanoma, until the benefit of retinol is confirmed in future well-designed prospective epidemiologic studies and clinical trials.Table S1Excluded publications after full-text review and the reasons for exclusion.(DOCX)Click here for additional data file.Checklist S1PRISMA checklist.(DOC)Click here for additional data file.


Genome-wide association studies have reported that the 9p21.3 locus confers risk for coronary artery disease (CAD). However, it is not known whether rs10811656 is linked with CAD in a Chinese population. Thus, the purpose of this study was to investigate the potential association between rs10811656 and the risk of CAD in a Chinese population.We conducted a hospital-based, case–control study with 251 CAD patients and 304 controls to examine the potential association of rs10811656 with CAD.P = 0.024). Compared to controls, rs10811656 was significantly associated with the stable angina pectoris , but not with acute coronary syndrome. There was also a highly significant association of rs10811656 with double-vessel and triple-vessel disease when patients were divided into subgroups based on the number of diseased vessels .The frequencies of the TT genotypes in CAD cases were significantly different from those in controls (adjusted OR: 1.96, 95 % CI: 1.09–3.505, Our results suggest that the rs10811656 locus might be associated with CAD in a Chinese Han population. Coronary artery disease (CAD) is one of the major causes of morbidity and mortality throughout the world, including China , 2. EpidThe 9p21.3 risk locus spans a >50-kb genomic region and comprises 59 linked single nucleotide polymorphisms (SNPs) that are in strong linkage disequilibrium. This genomic region encodes the cyclin-dependent kinase inhibitors, CDKN2A (p16 and p14) and CDKN2B (p15) , althougThe objective of the present hospital-based, case–control study was to investigate the potential association between rs10811656 and the risk of CAD, and further replicate the positive association of rs10811656 and ACS in a Chinese population.n = 304), who underwent coronary angiography to rule out CAD, were also enrolled during the same time period at the same hospital. All subjects enrolled in this study were of Han Chinese origin and had no history of significant concomitant diseases, including cardiomyopathy, bleeding disorders, renal failure, thyroid disease, pulmonary hypertension, and malignancy. This study was approved by the Ethics Committee of Shanghai Jiao Tong University Affiliated Sixth People’s Hospital, East Campus and informed consent was obtained from each participant. All authors had no access to information that could identify individual participants after data collection.This association study consecutively enrolled 251 unrelated consecutive CAD patients, who were all admitted to Shanghai Jiao Tong University Affiliated Sixth People’s Hospital, East Campus from January 2014 to December 2015. CAD was defined as ≥ 50 % luminal narrowing in at least one coronary artery. Patients with CAD were divided into subgroups with 1-vessel, 2-vessel or 3-vessel disease according to the number of significantly affected vessels using the Coronary Artery Surgery Study classification . Two carPeripheral venous blood was drawn from each participant in the first day after admission. Genomic DNA was extracted using the AxyPrep DNA Blood kit . The rs10811656 polymorphism was genotyped by PCR-LDR sequencing method, as reported previously .χ2) tests. Among the controls, the genotype frequencies were tested using the χ2-test test for the Hardy-Weinberg equilibrium (HWE). The association between the genotypes and CAD was evaluated by computing odds ratios (ORs) and 95 % confidence intervals (95 % CIs). A multiple logistic regression model was used to adjust for covariates including age, sex, body mass index, smoking status, hypertension, diabetes, and dyslipidemia. The linear trend in the association of rs10811656 polymorphism with CAD severity was evaluated by the χ2-test for trend. Two-tailed P values < 0.05 were considered significant. All of the statistical analyses were performed using SPSS for Windows version 13.0 .The allele frequencies and genotype distributions of the CAD cases and controls were compared using chi-square (The baseline characteristics of the case and control subjects are presented in Table P > 0.05). The distribution of the T genotype was also markedly different between CAD patients and controls (P = 0.024), as shown in Table P = 0.057). We also found that homozygous TT carriers had a 96 % increase in risk of CAD compared with CC carriers . In the subgroup of patients with stable angina pectoris (SAP), carriers of TT homozygous carriers had a 42 % increase in the risk of CAD compared with the CC carriers . However, no significant association was observed with ACS  frequency of rs10811656 was 0.3817, and the genotype distribution in our study subjects showed no deviation from the Hardy–Weinberg equilibrium  and smokers , but not in females  or non-smokers . Patients with CAD were subclassified into 3 subgroups  based on the number of stenotic coronary arteries. The frequency of patients with T variant genotypes increased from double- to triple-vessel disease, but not with single-vessel disease  and CDKN2B (p15), are known as suppressors that control cell cycle and cellular senescence .Harismendy et al. demonstrated that the 9p21.3 locus alters CAD risk by modulating inflammatory signaling through altered STAT1 binding and may modify immune responses by regulating expression of IFN, a 21 or related type I IFN; and this in turn affects the expression of the cyclin-dependent kinase inhibitor gene, CDKN2A/B, which is important regulator of cell proliferation, aging, and apoptosis . FurtherP = 0.057). We also found that homozygous TT carriers had a 96 % increase in the risk of CAD compared with the CC carriers (P = 0.024). In addition, TT genotype carriers had a 42 % increased risk of SAP ; however, the risk of ACS was not significantly increased. Our results are not consistent with the observations of Qiang et al. [In the present study, we observed that the distribution of T genotype was also markedly different between CAD patients and controls After adjusting for risk factors including age, sex, smoking status, hypertension, diabetes, and dyslipidemia, the OR for CAD in subjects with the variant genotypes (CT and TT) was 1.43, which was of borderline significance  had a significantly higher risk of CAD than the CC carriers. The T allele could be a genetic susceptibility marker for CAD in males. Valentina et al., reported that in males with breast cancer, the rs1011970/9p21.3 risk genotype was associated with HER2+ disease . Lin et We noted that among smokers, carriers of homozygous TT and heterozygous TC had an increased risk of CAD than CC genotype carriers. Our result suggests that the rs10811656 polymorphism may be factor that contributes to smoking-related CAD in a Chinese population. There was also no significant association between the rs10811656 genotypes and serum triglycerides, HDL-C and LDL-C levels in the CAD patients and controls. Therefore, our study revealed that the risk allele was associated with angiographically characterized coronary atherosclerosis, suggesting that the risk SNP exerts cell or tissue-specific effects relevant to atherosclerosis rather than affecting the traditional risk factors for CAD.Our study has several limitations. First, selection bias and a relatively small sample size in the present study may have reduced the statistical power of the results, even though the genotype distribution of patients and controls in our study was compatible with the Hardy–Weinberg expectations. Secondly, a prospective long-term outcome study is warranted to analyze the association of polymorphism with long-term prognosis in CAD patients. Lastly, our data should be extrapolated to other regions and ethnic groups cautiously.To the best of our knowledge, the present study demonstrates for the first time that the rs10811656 polymorphism is associated with increased risk of CAD in a Chinese population. Additionally, in male subjects and smokers, rs10811656 homozygous carriers of TT have an increased risk of CAD. Moreover, we found that the rs10811656 T variant is significantly correlated with SAP and angiographically characterized coronary disease severity. In order to enhance understanding of the relationship between this polymorphism and risk of CAD, further large-scale studies are required in other populations.