GSTM1, GSTT1, GSTP1, SULT1A1 and UGT1A1 genes and breast cancer clinicopathological characteristics together with disease progression.Breast cancer is the most frequent oncological disease among women. Estrogens are known to play an important role in breast cancer development. Recognition of the relationship between polymorphisms within estrogen metabolizing genes and conventional prognostic factors of breast cancer might improve our knowledge on individualized breast cancer prognosis. Therefore, we aimed to investigate possible associations between germline genetic polymorphisms within GSTP1 and SULT1A1 genotypes. GSTM1 and GSTT1 null genotypes were detected by multiplex PCR. UGT1A1 polymorphism was investigated with microsatellite analysis. Relationships between genotypes and breast cancer clinicopathological features along with disease progression were estimated by Pearson‘s Chi-square test. Logistic regression analyses were performed to estimate the odds ratios associating different genotypes with clinicopathological characteristics and disease progression.Our study involved 80 young (younger than 50 years of age) breast cancer patients. PCR-based Restriction Fragment Length Polymorphism (RFLP) assay was used to determine GSTT1 null genotype to have approximately 3.5 times higher risk for breast cancer progression than those with wild type genotype . Moreover, SULT1A1 G638A AA genotype significantly increased the chances of HER2 molecular subtype breast cancer when compared to GG genotype . Heterozygotes for GSTP1 A313G genotype were more likely to have positive lymph nodes in comparison to AA genotype carriers . No significant correlation was determined for UGT1A1 A(TA)nTAA and GSTM1 +/- polymorphism alone or combined GTTT1 null and GSTM1 null genotype.The study showed individuals with GSTT1 null genotype and SULT1A1 G638A AA genotype could be uselful genetic markers for breast cancer prognosis. Further analyses on larger sample size are required to highlight the effect of GSTP1 G allele on breast cancer prognosis.Conclusively, our findings suggest that The online version of this article (doi:10.1186/s12881-015-0147-4) contains supplementary material, which is available to authorized users. Breast cancer (BC) is the most frequent oncological disease among women. It is widely accepted that prolonged exposure to estrogens and their oxidative metabolites play an important role in BC developement. The carcinogenic effect of estrogens is thought to manifest in part through the ability of intermediate estrogen metabolites to form superoxide radicals and depurinating adducts that damage DNA ,2. GenetGSTM1, GSTT1, GSTP1, SULT1A1 and UGT1A1 encode for enzymes which have the ability to inactivate estrogen metabolites and remove them from the target tissues. The efficiency of estrogen metabolism depends on interindividual genetic variations, polymorphisms, within these genes.Genes GSTM1, GSTT1 and GSTP1 have been the most commonly studied.Glutathione S-transferases (GSTs) are a superfamily of enzymes which catalyze the conjugation of glutathione to a variety of chemicals, including estrogen metabolites, making them more water-soluble and easier to excrete. In such a way they decrease oxidative stress generated during estrogen metabolism. There are seven human cytosolic GSTs designated GST Alpha, Mu, Pi, Sigma, Omega, Theta, and Zeta . GSTM1, GSTM1 gene includes a deletion polymorphism which frequently affects both alleles. Deletion polymorphism in homozygous state (GSTM1 null) results in the absence of the enzyme. GSTT1 gene also has a deletion polymorphism. Homozygotes for the null allele of GSTT1 lack the respective enzyme. The GSTP1 gene contains several polymorphisms, including adenine to guanine transition at nucleotide 313 (A313G) which results in isoleucine to valine substitution in codon 105. The substitution is located in close proximity to the hydrophobic substrate-binding site and was shown to decrease the enzyme’s activity and affinity for electrophilic substrates but not for glutathione [tathione .SULT1A1 is Arg213His resulting from G to A transition at nucleotide 638. It was shown the His amino acid is associated with lower enzyme activity and lower stability than the wild-type amino acid [Sulfo-transferases (SULTs) are also involved in inactivation of estrogens by sulfating them to water-soluble metabolites. The most common polymorphism of ino acid .UGT1A1 is a dinucleotide repeat in the promoter region of the gene. The wild type allele consists of six TA repeats while the variant allele has seven repeats in the A(TA)nTAA motif. The seven repeat allele is known to be associated with decreased gene expression in comparison to the wild type allele [5 and (TA)8 alleles are rare.UDP-glucuronosyltransferases (UGTs) catalyze the conjugation of glucuronic acid to a variety of compounds including estrogens and their metabolites. One of the most common genetic variant of e allele . The  and conducted at the Hospital of Lithuanian University of Health Sciences, Kaunas Clinics. Informed consent was obtained from every participant.A group of 80 young (≤50 years of age) premenopausal female patients, diagnosed with stage I-II BC was collected. Women, who had additional cancer of any other location, were excluded from the study. Patient enrolment started in 2005 and follow-up was complete until 2012 12 31Clinicopathological data was collected from medical records for the analysis of differences among genotypes. Tumor estrogen receptor (ER), progesterone receptor (PR) and human epidermal growth factor receptor 2 (HER2) status had been determined immunohistochemically. HER2 was considered positive when imunohistochemically detected HER2 (3+) or (2+) amplification status was confirmed by positive Silver in Situ Hybridization. Tumour receptor status determined the choice of adjuvant therapy for clinicians according to approved prognostic and predictive breast cancer factors and following guidelines for breast cancer treatment at the time.Most of the studied BC patients (85%) had negative stromal lymphocyte infiltration. Almost half of the cases (48.8%) had positive lymph node involvement. Majority of the tumours (68.8%) were well to moderately differentiated (G1 or G2). Approximately half of the studied BC patients were positive for estrogen (51.2%) and progesterone (46.3%) receptors, while HER2 overexpression was determined only in 22.5% of tumors. The majority of the patients (47.5%) had luminal A tumor subtype. Triple negative, luminal B, HER2 molecular subtypes were observed in 30%, 12.5% and 10% of cases, respectively. Most of the tumours (63.8%) were not larger than 2 cm. During a follow-up period (of at least 2 years) 27.5% of patients were documented with a disease progression. Local and systemic disease spread  was considered to be disease progression.Genomic DNA was isolated from peripheral blood leukocytes using a commercially available DNA extraction kit  utilizing silica-based membrane technology.GSTT1 and GSTM1 genes were detected by multiplex polymerase chain reaction (PCR). The reaction conditions were described by Altayli and colleagues [ALB) was used as internal positive control. Primer sequences  method. DNA fragments were amplified for 37 cycles with annealing at 63°C. Primer sequences  of the second genotyping were in agreement with the ones obtained previously.GSTM1, GSTT1, GSTP1, SULT1A1 and UGT1A1 genotypes with clinicopathological characteristics and disease progression. P < 0.05 was considered statistically significant. Association analyses and logistic regression were carried out using SPSS  version 20.0 statistical software . Statistical power was also calculated when a statistically significant association was found.Prior to association analyses, deviation from Hardy-Weinberg equilibrium (HWE) was tested for each polymorphism using a Chi-square test. Relationships between genotypes and clinicopathological features together with disease progression were estimated by Pearson‘s Chi-square test. In cases where >25% of cells had expected value less than 5, Monte Carlo P values were assessed. Logistic regression analyses were performed to estimate the odds ratios (OR) associating different GSTM1 null and 17.5% were GSTT1 null. The frequencies of GSTP1 genotypes were as follows: 52.5% AA, 35.0% AG and 12.5% GG. Regarding SULT1A1 43.8% of cases were GG, 46.3% were AG and 10% were AA. UGT1A1 genotyping identified 43.8% 6/6, 42.5% 6/7 and 13.8% of patients to have 7/7 genotype. The genotypes of GSTP1, SULT1A1 and UGT1A1 were under HWE. Concerning GSTM1 and GSTT1, the genotyping assay used in this study does not distinquish homozygous wild type from heterozygous individuals, which makes it impossible to evaluate deviation from HWE.In a group of 80 patients studied 48.8% were We aimed to determine the linkage between the genotypes of genes involved in estrogen metabolism and clinicopathological characteristics of BC together with disease progression. The results . We also determined a significant association between SULT1A1 G638A genotype and HER2 molecular subtype of BC (P = 0.016). As far as GSTM1, GSTP1 or UGT1A1 genes are concerned  was also analyzed, however no statistically significant correlation was determined.There was a statistically significant correlation between SULT1A1 allele A to be significantly associated with HER2 molecular subtype only in a recessive model (P = 0.006). As our detection method for GSTM1 and GSTT1 deletions lack the information on heterozygosity, assessing the models for those polymorphisms was impossible. Other correlations were non-significant.Additional analyses were performed to clarify further the effect of low activity variant allele on BC phenotype and disease progression. The dominant and recessive models of each polymorphism are given in Table S3 . The association remained statistically significant after adjustment for all the treatments the patient received. The prevalence of HER2 BC subtype was significantly higher in individuals with SULT1A1 AA genotype  when compared to GG genotype. A allele significantly increased the chances of HER2 molecular subtype tumours only in a recessive model  but not in a dominant one (P = 0.099).Table GSTP1 AG genotype significantly increased the odds of positive lymph nodes in comparison to AA genotype , although no significant association was detected by Pearson Chi-square test.In addition, logistic regression analysis showed that In each case of a significant association, statistical power was calculated and varied between 59.5-97.1%.GSTM1, GSTT1, GSTP1, SULT1A1, UGT1A1 genotypes and BC phenotype together with disease progression. Our study found significant associations between polymorphisms in BC susceptibility genes and clinicopathological features of BC, as well as disease progression, with corresponding statistical power between 59.5-97.1%. These findings could provide useful prognostic information.To our knowledge this is the first study investigating the association between genotypes in five major estrogen metabolizing genes and BC characteristics. We assessed the relationships between GSTT1 null allele and disease progression, which remained statistically significant after adjustment for the treatments the patient received (P = 0.025). However, other associations of GSTT1 and tumor characteristics were non-significant. Pongtheerat and colleagues [GSTT1 deletion and lymph node involvement, status of estrogen, progesterone and HER2 receptors, and tumor size. This study was performed in Thai BC patients and no significant association was reported.We identified a statistically significant correlation between lleagues  also anaGSTP1 A313G genotype increases the chances of lymph node involvement if compared with AA genotype. Interestingly, homozygous GG genotype was not significantly related to lymph node status in our study. Bearing in mind the G allele is the low activity allele we anticipated the homozygous variant genotype to associate with positive lymph nodes. The absence of significant association might be due to a small number of individuals with GG genotype in our study. After all, Nedelcheva Kristensen et al. [versus AA). However, Pongtheerat et al. [et al. [GSTP1 A313G polymorphism and lymph node status in Thai and Spain BC populations, respectively. Prognostic value of axillary node status together with inconsistency between the data about G allele effect on lymph node metastasis points out to a necessity of larger studies in the future.Futhermore, our study showed that heterozygous t et al.  and Rome [et al.  reportedGSTP1 genotype other correlations were non-significant in our study, which is consistent with Romero and colleagues report [GSTP1 A313G polymorphism and PR status. Since PR expression has a prognotic value and is related to favourable prognosis, further investigation is needed to clarify the effect of GSTP1 on tumor characteristics.Regarding s report . Howevers report  identifiSULT1A1 increased the odds of HER2 molecular subtype of BC. HER2 amplification/overexpression is a strong prognostic factor for relapse and poor overall survival. HER2 molecular subtype BC is related to poor prognosis of the patients. Our finding that SULT1A1 AA genotype encoding the low activity enzyme is associated with a poor prognosis factor is in agreement with a general notion that a high sulfation capacity is protective against proliferative effects of estrogens.We revealed that AA genotype of SULT1A1 and clinicopathological BC characteristics were not detected in our study. Shatalova et al. [SULT1A1 genotype and BC phenotype, comprising tumour size and lymph node metastasis in BC patients of Russian ancestry. However, Han et al. [Other associations between a et al.  also didn et al.  announceGSTM1 null deletion does not play an important role on BC phenotype and disease progression. The results are in agreement with Medeiros et al. [et al. [et al. [GSTM1 deletion. However, Nedelcheva Kristensen et al. [GSTM1 null and GSTT1 null genotype correlated with positive lymph node status. Our study did not determine any association of combined GSTM1 null and GSTT1 null genotype with BC phenotype or disease progression.According to our results s et al. , Romero  [et al.  and Liza [et al.  reports,n et al.  reportedUGT1A1 gene did not reveal any significant association with studied tumor characteristics. Shatalova et al. [UGT1A1 A(TA)nTAA polymorphism and tumour size and grade but did not find any linkage. However, another study by Shatalova and colleagues [Furthermore, a et al.  also invlleagues  reavealeGenerally, the observed associations of lack of/ low enzyme activity with poor BC prognosis in our study are in agreement with a perception that lower efficiency of estrogen metabolite elimination leads to a more intense estrogenic stimulation of cancer tissue growth. Nonetheless, our study has some limitations. One of the limitations is small sample size. Larger cohort would allow conducting more precise subgroup analyses by different BC characteristics. Short follow-up period of our patients is another limitation of the study.GSTT1 null genotype and SULT1A1 G638A AA genotype could be useful genetic markers for breast cancer prognosis. Further analyses on larger sample size are needed to highlight the effect of GSTP1 G allele on breast cancer prognosis.In conclusion, our findings suggest that


Therefore, this meta-analysis aims to clarify the effects of TCF4 polymorphisms on FED risk.Studies investigating the associations between TCF4 and the risk of FED. Relevant studies were selected through an extensive search of PubMed, EMBASE, and the Web of Science databases. Pooled odds ratio (OR) and 95 % confidence interval (CI) were calculated using the random-effects model.A meta-analysis was conducted to assess the association between four single nucleotide polymorphisms (SNPs) inTCF4 rs613872 polymorphism and FED risk in all the genetic models tested . Statistic evidence was also detected for a significant association between three other SNPs and the risk of FED.Thirteen studies were included in this systematic review and meta-analysis. The pooled results showed that there was a strong positive association between the TCF4 polymorphisms  and the risk of FED.This meta-analysis suggested a genetic association between four  Fuchs’ endothelial dystrophy (FED) is a familial, slowly progressive, and irreversible disorder affecting the corneal endothelial cell monolayer . It has collagen type 8 alpha-2 (COL8A2), the sodium borate cotransporter gene (SLC4A11), transcription factor 8 (TCF8), and transcription factor 4 (TCF4), have been implicated in the pathogenesis of FED [TCF4 variations have been considered to be associated with a number of diseases, such as FED, schizophrenia, and primary sclerosing cholangitis [TCF4 gene is located on chromosome 18q [Although the precise reason for FED remains unclear, recent studies have reported significant insights into the genetic basis of the disorder. To date, several gene mutations, such as s of FED –8. Of thlangitis . The TCFene SLC4A, transcrsome 18q .TCF4 polymorphisms have attracted a lot of attention. A previous genome-wide association study (GWAS) showed a significant relationship between FED and four genotyped single nucleotide polymorphisms (SNPs)  of the TCF4 gene [TCF4 polymorphisms are associated with FED [TCF4 rs613872 polymorphism was not associated with FED, but Kuot et al. [TCF4 rs613872 polymorphism was a risk factor for FED. Other SNPs of TCF4  were also analyzed in some FED genetic association studies.Recently, with FED . Howevert et al.  found thTCF4, which include four SNPs: rs17595731, rs613872, rs9954153, and rs2286812.Meta-analysis is a powerful statistical technique that is often used in combination with different studies; therefore, it draws a more comprehensive conclusion. With this in mind, we conducted a meta-analysis to summarize all the relevant evidence for an association between the risk of FED and genetic polymorphisms of http://www.ncbi.nlm.nih.gov/pubmed/), ISI Web of Science (www.webofknowledge.com), and EMBASE (http://www.embase.com) databases. Key search terms included  and  and . References cited in each eligible literature were manual checked until no further studies were found. If the overlapping patient population was included in several studies, the latest study was included. The final literature search was updated on April 05, 2015, with no restrictions as to publication year, language, or methodological filter.Literature searches were performed in PubMed  odds ratio (OR) and the corresponding 95 % confidence interval (CI) were provided; and (d) sufficient genotypic or allelic information was provided to estimate. Exclusion criteria were: (a) case-only studies, familial studies, or duplicate data; (b) abstracts, comments, letters, reviews, or editorial articles; (c) insufficient genotyping data.Studies included in this meta-analysis were required to meet the following inclusion criteria: (a) designed as nested case–control, case–control, or GWAS; (b) evaluated Two observers (L.D. and S.H.Y.) independently extracted the following information from included studies, using a standardized data extraction form: first author, year of publication, country of origin, population ethnicity, source of controls, numbers of genotyped cases and controls, and Hardy-Weinberg equilibrium (HWE) for the control group. If a study provided several risk estimates, the best adjusted estimate was extracted. Any disagreement was resolved by discussion or adjudicated with the involvement of a third reviewer (P.X.Y.). The quality of each study was assessed using the Newcastle-Ottawa Scale (NOS) . The NOSTCF4 polymorphisms and FED. Adjusted ORs and 95%CIs were used if they were reported; otherwise, the pooled ORs and 95 % CIs without adjustments were calculated for the following genotypic models for rs613872 SNP: allele (G vs. T); homozygote (GG vs. TT), heterozygote (GT vs. TT), dominant (GG/GT vs. TT), and recessive (GG vs. GT/TT). Due of the insufficient data for the other three TCF4 SNPs, the pooled ORs and the 95 % CIs were calculated using an additive model. Data were combined using a random effects model to achieve more conservative estimates. Statistical heterogeneity between the studies was evaluated using Cochran’s Q test and the I2 statistic. For the Q statistic, p < 0.05 was considered to indicate statistically significant heterogeneity [2 was also used to assess the heterogeneity in the meta-analysis, and heterogeneity was said to exist when I2 > 50 % [p value less than 0.05 was considered statistically significant in the test for overall effect. The analysis was conducted using the Stata software package .ORs and 95 % CIs were used to assess the strength of the associations between ogeneity . I2 was 2 > 50 % . To dete2 > 50 %  and to c2 > 50 % , 23–25. 2 > 50 % , 27. A pTCF4 rs613872 polymorphism and FED. Three studies reported the association between the TCF4 rs9954153 polymorphism and FED. Three studies reported the association between the TCF4 rs17595731 polymorphism and FED. Five studies reported the association between the TCF4 rs2286812 polymorphism and FED. The study selection process is detailed in Fig. The initial search yielded 868 potentially relevant studies. After the removal of duplicates through electronic databases, 634 studies remained. Based on titles and abstracts, 616 articles were excluded because of their apparent irrelevance. In total, 18 full-text articles were further assessed for eligibility. Of these, six articles were excluded for the following reasons: the article was a review (n = 1) ; the artTCF4 rs613872 polymorphism and FED risk in all the genetic models tested  after removing the study by Nanda et al. It is interesting to note that when one study that deviated from the HWE principle was deleted [TCF4 rs613872 polymorphism and FED risk under an additive model  is presented in Fig. To evaluate the influence of an individual data set on the pooled results, one study was deleted at a time. The estimates were not altered substantially when any single study was deleted, suggesting the high stability of the meta-analysis results for a et al.  was the  deleted , the resTCF4 polymorphisms contribute to the development of FED, this correlation also needs to be verified by a case control study. Therefore, we performed this meta-analysis to provide the highest level of evidence for the association between TCF4 polymorphisms and the risk of FED. With a total of 5370 participants, this meta-analysis focused on the association between TCF4 variants and the risk of FED, thirteen studies addressing one or more SNPs in TCF4  were included. The pooled results showed that a significant association was detected between FED risk and the rs17595731, rs613872, rs9954153, and rs2286812 SNPs. This s consistent with the findings presented in the GWAS publication [Despite the fact that the GWAS publication noted that genetic variations in TCF4 rs613872 polymorphism and the risk of FED, which might be due to the different ethnicity of the included study participants [Differences in ethnic groups may affect genetic predisposition to human diseases , 37. In icipants . AnotherTCF4 gene and FED risk is still unclear. Possible mechanisms for the association between the TCF4 gene and FED risk have been proposed. First, the protein produced by the TCF4 gene (E2-2) might participate in endothelium growth, proliferation, and differentiation [TCF4 gene mutation associated with FED results from changes in the expression of ZEB1 [TCF4 gene might heighten the risk of FED.To date, the pathomechanism underlying the association between the ntiation . Therefontiation . In addi of ZEB1 . Taken t2 = 63.0 %; p = 0.013), and the heterozygote model  for rs613872. We performed the sensitivity analysis by exclusion of one individual study each time and we found that the study by Nanda et al. [An important aim of meta-analysis is to determine the sources of heterogeneity. In this meta-analysis, substantial heterogeneity was observed in the dominant model  and the risk of FED. Despite these encouraging findings, the inherent limitations of the studies should be considered, and conclusions drawn from the pooled results should be interpreted with caution. In the future, more epidemiologic studies of other ethnicities with a well-designed and larger sample size are needed to further confirm our findings.In conclusion, this meta-analysis suggested a genetic association between four


The present study provides preliminary evidence to demonstrate that obesity is a significant risk factor for aggressive PCa and PCa-specific mortality. The low survival rates observed among obese males with PCa may be a likely explanation for this association.A number of epidemiological studies have suggested that obesity is associated, albeit inconsistently, with the incidence of prostate cancer (PCa). In order to provide a quantitative assessment of this association, the present study examined the correlation between obesity and the incidence and associated mortalities of PCa in an updated meta-analysis of cohort studies. The cohort studies were identified by searching the EMBASE and MEDLINE databases on January 1, 2014. The summary relative risks (RRs) with 95% confidence intervals (CIs) were calculated using random-effects models. In total, 17 studies, which included 3,569,926 individuals overall, were selected according to predefined inclusion criteria. Based upon the results of the random-effects models, obesity was not significantly correlated with the incidence of PCa . However, further analysis revealed that obesity was significantly correlated with an increased risk of aggressive PCa . Furthermore, an increased risk of PCa-associated mortality was significantly associated with obesity , without any heterogeneity between the studies (I Prostate cancer (PCa) is a heterogeneous disease. In 2012, there were an estimated 241,740 new cases and 28,170 associated mortalities, making the disease the second most commonly diagnosed cancer and the second leading cause of cancer-associated mortality in males .Obesity, which is generally measured by the body mass index (BMI), has become a worldwide public health problem. In total, >1 billion people are estimated to be overweight BMI, ≥25), and >300 million of these are considered obese  . Accumul, and >30The present systematic review was conducted according to the Meta-analysis of Observational Studies in Epidemiology guidelines . On JanuTwo investigators screened titles and abstracts according to set inclusion and exclusion criteria. The primary aim of the present study was to confirm whether a positive association between obesity and the incidence of PCa existed. The criteria for study inclusion was as follows: i) A prospective or retrospective cohort design; ii) an exposure of interest of weight or BMI at baseline and/or at the end of the follow-up; iii) an outcome of interest in PCa and/or fatal PCa; and iv) publication in English. The meta-analysis included studies that reported standardized forms of relative risk (RR), risk ratio, hazard ratio or odds ratio, those with estimates on confidence interval (CI), or those that used RRs to represent various effect estimates. Due to a decreased risk of selection bias, results from cohort studies, as opposed to case-control studies, were chosen.The studies that were excluded from the meta-analysis were those that consisted of a case-control design. In addition to these criteria, reviews, meeting abstracts, editorials and commentaries were also excluded. Other studies that were not relevant to the meta-analysis, such as non-analytical epidemiology studies, or those that had not considered obesity and PCa as primary exposures and outcomes, were also excluded. Furthermore, studies that had failed to report these estimates, or those with only univariate estimates, were excluded. In the event that there were multiple publications from the same study population, only those that had combined independent studies, examined weight or BMI as the main interest of exposure, or presented the largest number of cases or follow-up years were included.The data were extracted independently by two authors, and then cross-checked in order to reach a consensus. The following variables were recorded: Last name of the first author, publication year, country in which the study was performed, participant characteristics, sample size, measure of association, mean length of follow-up, variables adjusted for in the analysis and the risk estimates with corresponding 95% CIs. From each study, the RR estimate that was adjusted for the greatest number of potential confounders was extracted. The quality of each study was assessed using the nine-star Newcastle-Ottawa Scale (NOS) . The maxThe RR value was used to measure the association of interest. In the event that the RRs were reported separately for subgroups by different levels of BMI, the results of the subgroups  were combined, and a common RR was calculated for the main analysis using a fixed-effects model. The highest levels were defined as ‘obesity’, and an overall pooled RR was calculated for the main analysis using a random-effects model. In studies that included BMI values that were only divided into two open-ended categories, the RR and CI estimates were assumed for the higher BMI category.2 test and quantified using the I2 metric. The random-effects model could have lead to additional weight and wider CIs being given to smaller studies than the fixed-effects model, and so was employed when heterogeneity was present . In the subgroup analysis of aggressive disease (Gleason score between seven and 10), obesity was significantly correlated with an increased risk of aggressive PCa , without any heterogeneity between studies  .In total, 14 cohort studies were identified that had recorded results on obesity and the incidence of PCa. As shown in P=0.754) . OverallObesity is an increasing worldwide health concern, and an independent risk factor for various types of cancer, including colon, liver, pancreas, breast and kidney cancer ,29. Obes2=56.4%; P=0.005). However, following the analyses of several parameters of PCa aggressiveness and progression, obesity was revealed to be significantly associated with high Gleason scoring or clinically advanced cases of PCa. Obesity was significantly correlated with an increased risk of aggressive PCa , without any heterogeneity between studies . The association between obesity and aggressive PCa was supported, as obesity was biologically linked with advanced disease. At present, the following three proposed mechanisms exist which aim to explain the association between obesity and aggressive PCa: The insulin/insulin-like growth factor-1 axis, the action of sex hormones and adipokine signaling  among these studies, there was evidence of heterogeneity .A previous meta-analysis of prospective cohort studies conducted by Cao and Ma  evaluateThe present study was subject to unavoidable limitations . FirstlyBased upon the results of the present meta-analysis, obesity was identified as an additional risk factor for aggressive PCa. Furthermore, obesity was associated with a risk of PCa-specific mortality. These results, which link obesity and PCa, highlight the importance of considering BMI when screening, treating and monitoring cases of PCa. Further investigations are required in order to evaluate the effects of weight loss in patients with PCa.


However, the meta-analyses did not detect significant genetic association between rs53576 and close relationships . In conclusion, genetic variation in the rs53576 influences general sociality, which further implies that it is worthy to systematically examine whether the rs53576 is a valid genetic marker for socially related psychiatric disorders.A common variant in the oxytocin receptor gene (OXTR), rs53576, has been broadly linked to socially related personality traits and behaviors. However, the pattern of published results is inconsistent. Here, we performed a meta-analysis to comprehensively evaluate the association. The literature was searched for relevant studies and effect sizes between individuals homozygous for the G allele (GG) and individuals with A allele carriers (AA/AG). Specifically, two indices of sociality were evaluated independently: i) general sociality , i.e., how an individual responds to other people in general; and ii) close relationships , i.e., how an individual responds to individuals with closed connections (parent-child or romantic relationship). We found positive association between the rs53576 polymorphism and general sociality (Cohen’s  Social interaction is important to almost every aspect of human life. However, people differ in their degree of sociality. Whereas some people enjoy an active social life and easily interacting with others, some people avoid social interactions and have difficulty in dealing with people. Extreme examples of the latter condition include several socially related psychiatric disorders, including autism and social anxiety disorder. Understanding the molecular genetic basis of individual differences in sociality has received significant research interest of late ,2.Previous studies have implied that the oxytocin system plays an important role in human socially related personality traits and behaviors ,4, hereaDue to the preservation of the oxytocin neuropeptide across mammalian species and the heritability of sociality in humans , variatiUsing a meta-analysis approach, a recent study conducted by Bakermans-Kranenburg and van IJzendoorn attempted to evaluate the association between the rs53576 polymorphism and sociality and did not establish a statistically significant genetic association . HoweverIn this study, we reexamined the association between rs53576 polymorphisms and sociality in humans with a meta-analytic approach. Compared with the previously mentioned meta-analysis , our stuTo identify eligible studies for this meta-analysis, we searched the PubMed and ISI Web of Knowledge for all publications available up to May 1st, 2014 that examined the association between the OXTR rs53576 polymorphism and psychological traits. The following terms were included in the search: oxytocin receptor (or OXTR) and gene. In addition, we identified additional relevant studies from the references of initially identified studies and recent literature reviews.We first identified all candidate measures from studies that examined the rs53576 polymorphism and psychological traits. Then, among all the psychological measures, phenotypes relating: 1) general sociality ; 2) close relationship ; and 3) depression were selected and classified into three independent categories respectively.Data were extracted by the first and last author of the current manuscript independently and then cross-checked. In addition, when a sample consisted of more than one eligible phenotype, the phenotype that appeared most frequently in the literature was selected. If the relevant data were not reported in the article, the authors were contacted and requested for data releasing. 22 out of 27 authors released their data in an appropriate format. One data point  was extracted from one sample.For each sample, the following information was extracted: first author, year of publication, country and ethnicity of the participants, proportion of females, diagnostic status , and age. Reverse scoring was used as necessary to maintain consistent measurement scaling across studies.d was used as the effect size measure, representing the standardized mean difference. Because the heterogeneous nature of included studies , the weighted effect size and 95% confidence intervalFollowing the convention in most published studies, OXTR rs53576 genotypes were grouped according to the presence and absence of the A allele (GG vs. GA+AA) ,11,13,31Z scores were used to assess the statistical significance of the weighted effect size.(95% CI) were generated via a random-effects analysis model by DerSimonian-Laird estimators . In addiPublication bias was assessed by the Begg’s Test and the Funnel plot in order to assess the tendency to publish positive results rather than negative results. A sensitivity analysis was applied to examine whether the inclusion of a particular study had a significant effect on the overall weighted effect size. That is, studies were removed one at a time to examine whether the pooled effect size remained statistically significant or non-significant. Finally, to examine the potential influence of any moderator variables, we performed meta-regression analyses with the average age of the sample, proportion of female participants, and proportion of Caucasian participants in the sample as predictors. Because not all studies reported demographic information, not all samples were included in the moderation analyses.Our meta-analysis was performed according to two guidelines: the “Preferred Reporting Items for Systematic Reviews and Meta-Analyses” (PRISMA) statement  35] and and35] aQ = 40.08, p = .02). Critically, the random effects analysis indicated an association  such that G allele homozygotes had higher general sociality than A allele carriers (p = .54) , the association observed in the meta-analysis was considered unlikely to be accounted for by any single outlying sample. Finally, meta-regression analyses suggested that the association was not significantly affected by sex (number of studies included (k) = 24, p = .20), age , or ethnicity .Eighteen studies –31,37–44carriers . The Begp = .54) . Becausep = .70), we used a random-effects model in the following meta-analysis because the tasks measuring close relationships were subjectively disparate  (p = .50) indicated no evidence of publication bias . Finally, meta-regression analyses indicated that sex , age , and ethnicity  did not significantly influence the association.Next, we tested whether the rs53576 polymorphism influenced the quality of close relationship, a specific kind of sociality measurement. Ten studies ,41,45–49p = .64) . The Begion bias , and theIn short, the aforementioned analyses revealed that rs53576 polymorphism explains the variances in general sociality, but not the variances in close relationship. This finding appears to be different from the findings of another meta-analytical study by Bakermans-Kranenburg and van IJzendoorn, who reported that no genetic association between rs53576 and sociality . To explps > .10). Note that several studies simultaneously included measures of general sociality and intimate relationship , Z = 1.65, p = .10) and second  analyses yielded statistically non-significant results. Second, when only data on general sociality were included, a trending genetic association was detected . Third, when data on close relationships were included, there was no evidence of genetic association .First, when general sociality and close relationships were categorized as a single phenotype, the random effects model indicated no evidence of genetic association between all sociality measures ps > .. Note thtionship ,38,41. TTaken together, our meta-analysis included more empirical studies, which increased the statistical power of the meta-analysis and facilitated detection of the genetic association; however, the phenomenon of increased effect size when the data were separated according to general sociality and close relationships was maintained even when only literature included in the previous meta-analysis was evaluated.n = 2177). As expected, there was evidence of between-study heterogeneity . The random effects analysis (GG vs. GA+AA) indicated no evidence of association  (p = 0.74) indicated no evidence of publication bias (p = .70), age , and ethnicity  did not significantly influence the association. In summary, there was no evidence of an association between the polymorphism and depression. Therefore, the genetic association with general sociality was unlikely to be mediated by depression.As we noted previously, depression and lack of sociality are often related . Therefop = .32) . Begg’s ion bias , and metd = 0.11). In addition, we did not find an association between rs53576 polymorphism and depression, suggesting that its genetic association with general sociality was unlikely to be influenced by depression. On the other hand, somewhat surprisingly, we did not find significant difference in measures of close relationship between G allele homozygotes and A allele carriers. In summary, our study suggested that the rs53576 polymorphism in the OXTR predicts how an individual generally responds to other people, but the polymorphism might be unrelated to individual differences in close relationships .In this study, we examined the relationship between the polymorphism of a widely-investigated common variant (rs53576) in OXTR and sociality. The meta-analysis on the overall samples revealed that G allele homozygotes were generally more sociable than A allele carriers  . Second,In conclusion, our study provides clear evidence of a relationship between the rs53576 polymorphism and general sociality. In the recent decade, genetic investigation of endophenotypes offers great opportunity as an alternative to investigations of categorical disease phenotypes . It is pS1 File(DOCX)Click here for additional data file.S1 Table(DOC)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.


FABP2 (rs1799883) Ala54Thr gene polymorphism and type 2 diabetes mellitus risk (T2DM) in various populations, but their results have been inconsistent. To assess this relationship more precisely, A HuGE review and meta-analysis were performed. The PubMed and CNKI database was searched for case-control studies published up to April 2014. Data were extracted and pooled odds ratios (OR) with 95% confidence intervals (CI) were calculated. Ultimately, 13 studies, comprising 2020 T2DM cases and 2910 controls were included. Overall, for the Thr carriers (Ala/Thr and Thr/Thr) versus the wild-type homozygotes (Ala/Ala), the pooled OR was 1.18 , for Thr/Thr versus Ala/Ala the pooled OR was 1.17 (95% CI = 1.05–1.41 P = 0.087 for heterogeneity). In the stratified analysis by ethnicity, the significantly risks were found among Asians but not Caucasians. This meta-analysis suggests that the FABP2 (rs1799883) Ala54Thr polymorphisms are associated with increased susceptibility to T2DM risk among Asians but not Caucasians.Many studies have examined the association between the  Type 2 diabetes mellitus (T2DM) has elevated prevalence, morbidity and mortality rates and the social and economic repercussions of its chronic complications compromise both the quality of life and productivity of those affected, making it a serious public health problem ,2. MeanwCurrently, more and more studies confirmed that T2DM was determined by the combined effects of multiple genetic and environmental factors . The mosFABP2) genes play a key role in the absorption and intracellular transport of dietary long chain fatty acids. The FABP2 gene is located in the 4q28–4q31 chromosomal region, consists of ∼3.4 kbs [FABP2 is a member of the FABPs superfamily that produces intracellular proteins which bind hydrophobic ligands reversibly [Fatty acid-binding protein 2  Ala54Thr polymorphisms in T2DM. However, the results of these studies remain inconclusive. A single study might not be powered sufficiently to detect a small effect of the polymorphisms on T2DM, particularly in relatively small sample sizes. Various types of study populations might also have contributed to these disparate findings. To clarify the effect of the FABP2 (rs1799883) Ala54Thr polymorphism on the risk for T2DM, we performed a meta-analysis of all eligible case-control studies that have been published.Recently, many studies have investigated the role of the FABP2’ ‘polymorphism’ ‘type 2’ and ‘diabetes mellitus’. An upper date limit of 1 April 2014 was applied; no lower date limit was used. The search was performed without any restrictions on language and was focused on studies that had been conducted in humans. We also reviewed the Cochrane Library for relevant articles. Concurrently, the reference lists of reviews and retrieved articles were searched manually. Only full-text articles were included. When the same patient population appeared in several publications, only the most recent or complete study was included in this meta-analysis.We searched for studies in the PubMed and CNKI  electronic databases to include in this meta-analysis, using the terms ‘fatty acid-binding protein 2’, ‘i) evaluated FABP2 Ala54Thr polymorphisms and T2DM risk; (ii) were case-control studies; (iii) supplied the number of individual genotypes for the FABP2 Ala54Thr polymorphisms in T2DM cases and controls respectively; and (iv) demonstrated that the distribution of genotypes among controls were in Hardy–Weinberg equilibrium.For inclusion, the studies must have met the following criteria: they  versus the wild-type homozygotes (Ala/Ala), Thr/Thr versus Ala/Ala were calculated respectively. Subgroup analyses were done by ethnicity. Heterogeneity assumptions were assessed by chi-square-based Q-test [P-value greater than 0.10 for the Q-test indicated a lack of heterogeneity among the studies. Thus, the pooled OR estimate of each study was calculated by using the fixed-effects model (the Mantel–Haenszel method) [Odds ratios (ORs) with 95% CIs were used to determine the strength of association between the d Q-test . A P-val method) ; otherwi method) .One-way sensitivity analyses were performed to determine the stability of the results-each individual study in the meta-analysis was omitted to reflect the influence of the individual dataset on the pooled OR .t-test, as suggested by Egger  [Potential publication biases were estimated by funnel plot, in which the standard error of log (OR) of each study was plotted against its log (OR). An asymmetrical plot suggests a publication bias. Funnel plot asymmetry was assessed by Egger's linear regression test, a linear regression approach that measures the funnel plot asymmetry on a natural logarithm scale of the OR. The significance of the intercept was determined by on bias) .All calculations were performed with STATA, version 10.0 .A total of thirteen publications involving 2020 T2DM cases and 2910 controls met the inclusion criteria and were ultimately analysed –27. Tablversus the wild-type homozygotes (Ala/Ala), the pooled OR for all studies combined 2020 T2DM cases and 2910 controls was 1.18 .Table versus (Ala/Ala)  and Thr/Thr versus Ala/Ala . However, for Caucasians, significantly risks were not found for (Ala/Thr and Thr/Thr) versus  and Thr/Thr versus Ala/Ala .In the stratified analysis by ethnicity, significantly risks were found among Asians for (Ala/Thr and Thr/Thr) A single study involved in the meta-analysis was deleted each time to reflect the influence of the individual data set to the pooled ORs, and the corresponding pooled Ors were not materially altered (data not shown).versus (Ala/Ala) showed that the Egger test was not significant . These results did not indicate a potential for publication bias.Begg's funnel plot and Egger's test were performed to access the publication bias of literatures. Evaluation of publication bias for (Ala/Thr and Thr/Thr) FABP2 gene has been proposed as a candidate gene for diabetes and insulin resistance because the protein it encodes is involved in fatty acid absorption and metabolism [FABP2 gene variants and their association with insulin resistance and T2DM. The most studied variant is the Ala54Thr variation at codon 54, a mis-sense variant that has a definite effect on the primary structure of the protein and affects its fatty acid-binding properties. Previous studies have found a significant association between the FABP2 genotype and occurrence of T2DM or decreased insulin sensitivity [FABP2 Ala54Thr genotypes was shown to be associated with increased fatty acid-binding, increased fat oxidation, and insulin resistance [FABP2 gene did not influence insulin sensitivity.The tabolism . Numerousitivity . In Pimasistance . On the sistance , the AlaFABP2 Ala54Thr polymorphisms and T2DM risk based on a Human Genome Epidemiology review of all research published up to April, 2014. This meta-analysis explored the association between the FABP2 Ala54Thr polymorphisms and T2DM risk. Our results indicated that FABP2 Ala54Thr polymorphism was significantly associated with the susceptibility to T2DM among Asians but not in Caucasians.To our knowledge, the current meta-analysis firstly elucidated the potential association between the Population stratification is a troubling issue and can lead to spurious evidence on the association between biomarkers and a disease, implicating the disparate effects of environment and ethnic differences on genetic background . In thisIt is important to keep in mind that type 2 diabetes is clearly multifactorial in origin, being determined by both genetic and environmental factors. Body composition, habitual PA levels, and diet are factors that are at least partly environmentally determined, and all are known to exert their own substantial and independent effects on insulin resistance and type 2 diabetes. Environmental factors clearly must play a substantial role in determining type 2 diabetes prevalence. During this same time, factors such as habitual PA levels and body composition have changed substantially, while negligible genetic alterations could have occurred over this time frame. Multiplegene/environment interactions ,34 wouldet al. study [FABP2 Ala54Thr polymorphism and insulin resistance and blood glucose. They found that FABP2 Ala54Thr polymorphism was weakly associated with a higher degree of insulin resistance, higher level of fasting insulin and higher level of 2-h BG. However, they did not perform the meta-analysis on the association of the FABP2 Ala54Thr polymorphism with risk of type 2 diabetes because of few studies on this association. In addition, that meta-analysis did not perform subgroup analyses by ethnicity. Since then, several additional studies with large cohort populations have been reported. We have performed this meta-analysis by including more recent related studies and by generally by using a more comprehensive search strategy. Study screening and data extraction were performed independently and reproducibly by two reviewers. Moreover, we also explored the association between the FABP2 Ala54Thr polymorphisms and T2DM risk for subgroups divided according ethnicity stratification.Zhao l. study  performeSome limitations of this meta-analysis should be acknowledged. First, heterogeneity can interfere with the interpretation of the results of a meta-analysis. Although we minimized this likelihood by performing a careful search of published studies, by using explicit criteria for a study's inclusion and performing strict data extraction and analysis, significant interstudy heterogeneity nevertheless existed in nearly every comparison. The presence of heterogeneity can result from differences in the selection of controls, age distribution, and prevalence of lifestyle factors. Although most controls were selected from healthy populations, some studies had selected controls among friends or family members of lung cancer patients or patients with other diseases. Further, only published studies were included in this meta-analysis. The presence of publication bias indicates that non-significant or negative findings might be unpublished. Finally, our results were based on unadjusted estimates; a more precise analysis should have been conducted if individual data were available, which would have allowed us to adjust using other covariates, including age, ethnicity, family history, environmental factors and lifestyle.FABP2 Ala54Thr polymorphisms are associated with the increased susceptibility to T2DM risk among Asians but not Caucasians. Large-sample studies of different ethnic groups with carefully matched cases and controls should be considered in future association studies to confirm the results of our meta-analysis.Despite these limitations, this meta-analysis suggests that the


IFN-γ +874T/A (rs2430561) polymorphism were related to the risk of many types of cancer. However, the association between IFN-γ +874T/A polymorphism and leukemia risk remained controversial.Interferon gamma (IFN-γ) has antitumor and antiproliferative effects, and previous studies indicated We performed a comprehensive meta-analysis based on the Preferred Reporting Items for Systematic Reviews and Meta-analyses statement (PRISMA). Electronic database of Embase, Pubmed, and the Cochrane Library were searched for eligible articles published up to December 13, 2015. The association between genetic polymorphisms and leukemia risk was measured by odds ratios (ORs) and its corresponding 95% confidence intervals (CIs).IFN-γ +874T/A polymorphism and overall leukemia risks were lacking, decreased chronic lymphocytic leukemia (CLL) risk was detected in the allelic model , the codominant model , and dominant model  by using fixed-effect model separately. On the contrary, results indicated T carries have an increased chronic myelogenous leukemia (CML) risk in dominant model .A total of 8 studies amounting to 420 patients and 767 control subjects were retrieved for this study. Although associations between IFN-γ +874T/A polymorphism are related to CML and CLL risk. In addition, our work also points out IFN-γ +874T/A polymorphism may play dual contrasting role in leukemia risk.This study suggests  Cytokines and their genes play a fundamental role in the immunopathogenesis of malignant, infectious, and autoimmune disease.5 It is, therefore, important to identify these genes variants contributing to leukemia pathogenesis.Leukemia is one of the most common malignancy in human, with an estimated 368,800 new cases and 303,200 deaths occurred in 2012 worldwide.6 The production of IFN-γ is under genetic control. The SNP  of IFN-γ gene, a T-to-A transition at +874 loci in the 1st intron of IFN-γ gene, is complete linkage disequilibrium with the12 CA repeat microsatellite and correlated with IFN-γ production.7 Moreover, the transcription factor nuclear factor kappa-B (NF-κB) binds preferentially to the +874T allele, which is known to upregulate the expression of IFN-γ.8 The +874T/A polymorphism generates 3 genotypes , and TT genotype produces a high level of IFN-γ, helps the host's defense against viral infection. Conversely, the genotypes AA and AT cause relative low IFN-γ production, which may increase the risk of viral infection.8 It appears that the +874T/A polymorphism can influence IFN-γ expression, and in this way, affects immune response.Interferon gamma (IFN-γ) is a T-helper 1 (Th1) proinflammatory cytokine that plays a pivotal role in antiviral activities and has antitumor and antiproliferative effects.IFN-γ genes and leukemia was previously studied,9 in addition, a number of studies have attempted to assess associations between IFN-γ +874T/A polymorphism and leukemia risk.16 However, a unified conclusion is still unreached, furthermore, most of these studies had insufficient sample sizes of studied populations, which led to lower statistical power. Therefore, we conducted a meta-analysis to evaluate broadly the available evidence on the association between IFN-γ +874 T/A polymorphism and leukemia risk.The relationship between 17This meta-analysis was conducted according to Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement (S1_Table), including the search strategy, selection criteria, data extraction, and data analysis.Two investigators (YS and ZW) searched independently for all publications in Embase, PubMed, and Cochrane Library up to December 14, 2015, by using the following search terms: “interferon gamma,” “IFN-γ,” “rs2430561,” “polymorphism,” and “leukemia.” Articles in reference lists were also hand-searched. Only human studies were searched and there was no restriction on the language.Case–control or cohort design studies evaluated the associations between IFN-γ+874T/A polymorphism and leukemia risk, and the controls should be healthy populations.Studies offering the ability to extract data for calculating the odds ratio (OR), 95% confidence intervals (CIs), and Hardy–Weinberg equilibrium (HWE).The 2 investigators choose the suitable studies according to the following criteria:Review articles, letters, case reports, editorials, and conference abstracts.Family based studies.The articles did not include genotype frequencies, and this information was not available from the authors contacted.Duplicate data.The exclusion criteria were:The based characters and data included the first author's name, publication date, country, ethnicity, diagnosis criteria for leukemia, source of control, total sample size, genotyping method, and the genotype frequencies of cases and controls were extracted independently by the 2 investigators (YS and ZW) from the included studies. To determine the accuracy of the extracted information, the data extracted by the 2 investigators should be all the same, and if there was a dispute they should check their data again. If the 2 investigators could not reach an agreement, the dispute was submitted to a 3rd reviewer (CL) to decide. In addition, if the genotype frequencies of cases and controls could not be extracted directly from the article, the 2 investigators should try to contact the authors for help, or obtain the data from previous literatures. HWE was calculated according to the genotype frequencies of controls. The quality of the studies included in this meta-analysis was assessed according to HWE, and we considered a high quality study should be fall in HWE.The association between genetic polymorphisms and leukemia risk was measured by ORs and its corresponding 95% CIs. The allelic model (T vs A), codominant model , dominant model (TT + TA vs AA), and recessive model (TT vs TA + AA) were examined.Q-test and I2 statistics.19 The fixed-effects model (the Mantel–Haenszel method) is used when the effects are assumed to be homogenous (PQ ≥ 0.1 or I2 < 50%). Otherwise, the random effects model (the DerSimonian and Laird method) is used when they are heterogeneous (PQ < 0.1 or I2 ≥ 50%). Subgroup analysis was carried out stratified by leukemia type, ethnicity (Caucasian and Asian), and HWE . Begg's funnel plot and Egger's test were used to investigate the publication bias in the meta-analysis; P < 0.05 indicated that the result was statistically significant. Sensitivity analysis was performed by omitting 1 study at a time.Heterogeneity was assessed by a Chi-squared All the tests in this meta-analysis were conducted with RevMan 5.2.7 (Cochrane Collaboration) and STATA software .An ethical approval was not necessary needed as this is a meta-analysis based on previous studies.11 was separated as 2 studies for they evaluated 2 different diseases, therefore, a total of 8 studies amounting to 420 patients and 767 control subjects were retrieved for this study. Two studies populations come from Asian12 and 6 studies were Caucasian.16 Two studies were failed to fall in HWE.14The literature search and study selection procedures are shown in Figure IFN-γ +874A/T polymorphism, there was no significant association between IFN-γ +874A/T polymorphism and leukemia risk under all comparison models.The meta-analysis results for all the studies are shown in Table P = 0.002, I2 = 19.0% and P = 0.295 for heterogeneity, Figure P = 0.009, I2 = 0.0% and P = 0.863 for heterogeneity), the codominant model , and dominant model  among Asian group by using fixed-effect model.However, significant heterogeneity between studies was detected in comparison model expect recessive model. Therefore, we further performed subgroup analysis Table . In the I2 > 50%, Table Significant heterogeneity was discovered in this study expect the recessive model in the overall population. Then we carried out subgroup analysis according to leukemia type, and the results showed there was no obvious heterogeneity in the subgroup analyses of CLL and CML, expect TT vs TA analysis  have a decreased risk of CLL compared with AA genotype. On the contrary, results from dominant model indicated T carries have nearly 1.783-fold increased CML risk. To our best knowledge, this is the 1st evidence suggested IFN-γ +874T/A T allele increased cancer risk from meta-analysis.This is the 1st quantitative evaluation on IFN-γ is a protect genotype for CLL is biologically plausible. First, the +874 position, which is in strong linkage disequilibrium with the 12 CA repeat microsatellite in the noncoding region of the 1st intron of IFN-γ gene, controls the production of IFN-γ.20 As we known now, IFN-γ is a proinflammatory cytokine playing a pivotal role in both innate and adaptive immune responses, and has antitumor and antiproliferative effects.6 Therefore, it is understandable that T allele, which with high level IFN-γ production, may be a protect gene for CLL. Second, the DNA sequence containing the +874 T allele is the specific binding site for NF-κB.21 NF-κB knockout mice have been shown to have low values of IFN-γ and an increased susceptibility to infection.22 This polymorphism leading to the dysfunction of NF-κB pathway may cause oxidative damage, which also can increase the risk of cancer.21 Indeed, our finding is consistent with previous meta-analysis that T allele is a protect genotype among patients in breast cancer, cervical cancer, asthma, and leprosy.26T allele in the +874 position of 27 and this is also not a new concept for the roles of cytokines in cancer. Transforming growth factor (TGF)-β has an ability to induce malignant behavior of normal fibroblasts, at the same time, it has profound growth-suppressive effects on many cells.28 Tumor necrosis factor (TNF), an extraordinarily pleiotropic cytokine, also displays pro- and antitumoral effects.29 A potential mechanism is localized inflammation may lead to an immunosuppressive and tolerogenic tumor microenvironment.30 For the dark side of IFN-γ, Zaidi and Merlino31 believed it may lie in the homeostatic functions of IFN-γ. Although high level of IFN-γ can protect normal cells from the collateral damage, these same mechanisms may also allow cancer cells to evade attack from the immune system.31 NK cells play a central role in tissue repair, however, Taniguchi et al32 found IFN-γ can enhanced resistance to NK cells, which was confirmed by another study and possible mechanism is higher metastatic ability of IFN-γ clones attributed to increased resistance to NK cells.33 But evidence so far for the roles of IFN-γ that may be more protumorigenic has been indirect, the exact mechanisms might IFN-γ behave as a “bad guy” in cancer need to be confirmed by further studies.However, this study observed T allele is a risk genotype for CML. These results suggested 2 faces of IFN-γ in leukemia. Indeed, T allele is a risk genotype for diseases had been observed in bipolar disorder,34 Meanwhile, the predisposition of every cancer types could be differentially influenced by genetic factors. Previous studies have showed the association between IFN-γ +874T/A polymorphism and cervical cancer,36 bladder cancer,37 liver cancer,38 and breast cancer,39 and suggested that T allele might be a protective genotype in cancer for TT genotype produces a high level of IFN-γ. Yet a recent meta-analysis showed no association between IFN-γ +874T/A polymorphism and cancer susceptibility,40 at the same time, the other meta-analysis found positive association in cervical cancer and breast cancer,26 which suggested that IFN-γ +874T/A polymorphism play different roles in different cancer types; therefore, the conclusion from a pooled analysis for association between IFN-γ +874T/A polymorphism and all cancer types may be unreliable. This meta-analysis confirmed this point. We found no significant association between IFN-γ +874A/T polymorphism and leukemia risk when all leukemia types were pooled into the meta-analysis, however, when stratified by leukemia types, results suggested T allele contributed completely opposite influence between CML and CLL. These results not only could partly explain the inconsistent results in different leukemia types but are also the reason why we conducted this study.Genetic susceptibility is believed to be an important factor in the development of cancers.IFN-γ +874T/A were quite different among different regions and ethnicities.4 However, we noticed that there are 2 studies on CLL and 3 on CML in Caucasians, which may lead to negative results happened in the pooled analysis considering the dual role of IFN-γ +874T/A in CLL and CML.This study performed a subgroup analysis stratified by ethnicities and the significant associations were only found in the dominant model among Asians. Therefore, the genetic background may be another important factor contributed to the inconsistent results. Indeed, genetic background is distinct among different populations and our previous study had shown the genotype and allele frequencies of 43 Then, Berndt et al44 discovered 10 independent SNPs in 9 novel loci in 2013 and Speedy et al45 identified five SNPs in new susceptibility loci in 2014 by GWAS separately. These studies suggest that additional risk loci should be identified by more studies. Our results suggested that rs2430561 in IFN-γ on chromosome 12q24 contributes to susceptibility to CML and CLL based on traditional genotyping methods, and these results have motivated us to carry out a GWAS focus on the SNPs on chromosome 12 to discover novel susceptibility loci for leukemia in the future.Genome-wide association studies (GWAS) have successfully identified 13 loci associated with risk of CLL/small lymphocytic lymphoma before 2011.10 which can easily lead to false positive or negative results. Thus, we need to be taken into consideration when interpreting the findings. Further study with larger sample numbers will be helpful to improve the power for detecting the positive effects. In addition, other confounders, such as age and gender, were not adjusted because it is hard to get original data from author. In spite of these, our meta-analysis had some advantages. In the first place, this study greatly increased the statistical power compared with individual studies. There is one more point, no significant heterogeneity was detected in subgroup analysis stratified by leukemia types, together with publication bias and sensitivity analysis results indicated that our results were moderately robust.In clinical practice, patients with leukemia are much less than that of lung cancer, liver cancer, breast cancer, and cervical cancer; therefore, the most limitation of this meta-analysis is that the present studies on susceptibility of leukemia are all with relative small sample sizes. As an example, we noticed the TA genotype frequency is only 1 in the study by Amirzargar et al,IFN-γ +874T/A polymorphism contribute to CML and CLL risk. However, our results are still needed to strengthen by further studies with larger sample sizes.Taken together, results from this meta-analysis demonstrate that


Increased ratio of n-3/n-6 polyunsaturated fatty acids (PUFAs) in diet or serum may have a protective effect on the risk of breast cancer (BC); however, the conclusions from prospective studies are still controversial. The purpose of this study is to ascertain the relationship between intake ratio of n-3/n-6 PUFAs and the risk of BC, and estimate the potential summarized dose–response trend.Relevant English-language studies were identified through Cochrane Library, PubMed and EMBASE database till April 2013. Eligible prospective studies reporting the multivariate adjusted risk ratios (RRs) for association of n-3/n-6 PUFAs ratio in diet or serum with BC risk. Data extraction was conducted independently by 2 investigators; disagreements were reconciled by consensus. Study quality was assessed using the Newcastle-Ottawa scale. Study-specific RRs were combined via a random-effects model.P for linear trend = 0.012). USA subjects with higher ratio of n-3/n-6 in serum phospholipids (PL) have a significantly lower risk of BC , and per 1/10 increment of ratio in serum PL was associated with 27% reduction of BC risk .Six prospective nested case–control and 5 cohort studies, involving 8,331 BC events from 274,135 adult females across different countries, were included in present study. Subjects with higher dietary intake ratio of n-3/n-6 PUFAs have a significantly lower risk of BC among study populations , and per 1/10 increment of ratio in diet was associated with a 6% reduction of BC risk (pooled RR = 0.94; 95% CI: 0.90, 0.99; Higher intake ratio of n-3/n-6 PUFAs is associated with lower risk of BC among females, which implies an important evidence for BC prevention and treatment is by increasing dietary intake ratio of n-3/n-6 PUFA. No firm conclusions from USA populations could be obtained, due to the limited numbers of USA studies. The lOR) etc.-9. ThereThe studies from cell lines and animals have shown promising results of down regulating BC tumor growth by n-3 PUFAs as a nutrient to compete with n-6 PUFAs,11. Most“Fatty Acids, Omega-3″OR “Fatty Acids, Omega-6″) AND “Breast Neoplasms” for PubMed, “Breast tumor” AND (“omega 3 fatty acid” OR “omega 6 fatty acid”) for EMBASE and “Fatty Acids”AND “Breast Neoplasms”  for Cochrane Library databases. We also searched systematic reviews from the above-mentioned database, and checked reference lists to identify studies that might have been missed. The present meta-analysis was conducted using the standard methods from Cochrane Collaboration, and reporting items were mainly based on MOOSE guidelines for meta-analysis of observational studies[We identified prospective studies which reported the association between intake ratio of n-3/n-6 (n-6/n-3) and BC risk up until April 2013 from PubMed, Embase, and Cochrane Library database using literature retrieval of subject headings. Search strategy was ( studies (Additio1) Participants: study population included any adult women , whose base conditions were regarded as stable; 2) Exposure: evaluating ratio of n-3/n-6 (n-6/n-3) PUFAs in diet or human serum (plasma) PL; 3) Outcomes: evaluating BC incidence as outcome variable and providing risk ratios (RRs) for all categories of dietary or serum (plasma) PL ratio of n-3/n-6 or n-6/n-3 PUFAs; 4) Study Design: prospective studies  were included.Two trained investigators (YF and JG) identified articles eligible for further review by performing a stepwise screening of titles or abstracts, followed by a full-text review based on common inclusion criterion. Discrepancies were resolved through discussion with the third investigator (BY). Studies of cross-sectional, cross-over, randomized controlled trials (RCT), experimental designs , non- original research , abstract, unpublished studies, or duplicated studies were excluded. Our search was restricted to human studies published in English. We did not contact authors for the detailed information of primary studies only reporting association of n-3 or n-6 PUFA with BC risk. We contacted the authors of the two studies reporting association of n-3 and n-6 PUFA with BC risk by email.Data extraction was finished independently and performed twice by two reviewers (YF and JG), and disagreements were reconciled by consensus. Detailed data concerning participants, exposure, comparability and outcomes were extracted using a standard extraction form  to total n-6 PUFAs in diet or serum (plasma) PL, and pooled RR including corresponding 95% CI was taken as the summary risk estimate for all studies. RRs from each study were firstly transformed to their logarithm (logRR), and corresponding 95% CIs were used to calculate corresponding standard errors (selogRR). We conducted two types of meta-analysis. Firstly, we conducted meta-analysis for the highest quantile  compared with lowest or reference, and study-specific RRs were combined using a random-effects model described by DerSimonian and Laird, which cQ test and I2 statistic. We considered an I2 value greater than 50% and 2-tailed P < 0.10 as indicative of heterogeneity according to Cochrane Handbook, and defined the low, moderate and high degrees of heterogeneity by I2 values of 25%, 50% and 75% as cut-off points[Heterogeneity was assessed with the f points respectiP < 0.05 was considered representative of statistical significance)[Sensitivity analysis was performed to evaluate potential influence of individual study on overall risk estimation, and compare the pooled RR from random effect model with that from fixed effect model. Potential publication bias was qualitatively delineated by the asymmetry of funnel plot, and it was also quantitatively examined by Begg’s test and Egger’s regression test (ficance). If poteficance). This meficance). StatistWe identified 1,112 potential studies from electronic search, and 772 studies were left after removing duplicates. Thirty one prospective studies were obtained after title and abstract review. Eleven studies were eligible for inclusion in the present study after full text review Figure .Included studies consist of 5 prospective cohort studies-17,32,33I2 = 11.40%; Pheterogeneity = 0.33).We performed a random-effects model meta-analysis for highest quantile compared with lowest Figure . Eleven Plinear = 0.012; Pcurvilinear = 0.018) among study populations. Per 1/10 increment of dietary n-3/n-6 ratio was associated with a 6% reduction of BC risk   among study populations, and the pooled RR for per 1/10 increment of n-3/n-6 ratio in serum PL was 0.89  , and per 1/10 increment of n-3/n-6 ratio in serum PL was associated with a 27% reduction of BC risk . There was no statistical significance of dose–response trend among Europe populations from 2 studies , and the pooled RR of per 1/10 increment of ratio in serum PL was 1.03  .Four articles reporting PUFAs as compositions were eligible for the dose–response association of serum PL biomarker with BC risk,19,33-36I2 = 0.00%) and 59,125 participants form 5 European studies , results of metaregression did not show a significant difference between the three populations. When stratified by the adjustment for potential confounders, metaregression did not show a significant difference between estimates adjusted and those not adjusted for specific covariates.Metaregression and subgroup analysis was performed to explore the probable source of heterogeneity Table . AlthougP for metaregression = 0.870; P for a permutation test = 0.990).When separately analyzed by exposure assessment, we found a significantly negative association between BC risk and n-3/n-6 ratio in diet from 5 prospective cohort and 1 nested case–control studies, involving 7,385 BC invents and 271,357 participants , whereas no significant association was found between BC risk and serum PL biomarker from 5 prospective nested case–control studies, involving 946 BC cases and 1832 controls . However, there was no apparent difference between studies using diet ratio of n-3/n-6 as exposure and those using serum biomarker of n-3/n-6 as exposure with metaregression (I2 = 0.00%) after excluding the study by Chajes[Finally, we conducted sensitivity analysis and publication bias analysis Table . For diey Chajes, indicaty Chajes.P = 0.244) and Egger’s regression test (P = 0.138) showed no evidence of possible publication bias. Contour-enhanced funnel plots of random effect model showed studies appear to be missing in areas of high statistical significance (p < 0.05), indicating that publication bias is a less likely cause of the funnel asymmetry . Fatty acid profile in serum phospholipids may be an untypical representative of sensitive biomarkers indicating post-absorptive amounts and change at the target tissue, which did not provide the same information as the dietary questionnaire tools. Therefore, the strength of the relationship between the ratio in serum PL and BC risk might be lower compared with the relationship between the ratio in diet and the risk of BC.The present study, involving 8,331 BC events from 274,135 participants, indicated that higher ratio of n-3/n-6 PUFAs is associated with lower risk of BC, and per 1/10 increment of n-3/n-6 ratio in diet is associated with a 6% reduction of BC risk. However, subgroup analysis showed there was no significant relationship between serum PL biomarker and BC risk, but the significant difference between association of dietary n-3/n-6 ratio and serum PL with BC risk was not observed with metaregression (I2 = 21.00%) in this meta-analysis indicated the subtotal variation was attributable to between individual studies. There are some possible reasons for explaining the potential heterogeneity in present study. Firstly, the heterogeneity from geographic regions could be partially explained by different dietary patterns. The n-3/n-6 ratio in Japanese diet is about 1 to 4, whereas that in western diet (USA and Europe) is about 1 to 15–20[In vivo, AA  and EPA  can simultaneously compete for the same cyclooxygenases (COX) and lipoxygenases (LOX) metabolism pathways, leading to the production of n-6 family derived 2-series PG and 4-series LT with promoting tumor growth effects[3 (PGE3)/prostaglandin E2 (PGE2) to inactivate the activity of adipose aromatase P 450 which catalyzes the conversion of 19-carbon steroids to estrogens, and thus reduce estrogen-stimulated cell growth action[Although the was no significant difference between most subgroups with metaregression, study heterogeneity , and exclusion of any individual study did not significantly change the negative relationship between intake ratio of n-3/n-6 PUFAs and BC risk. Although publication bias could be of concern because small studies with null results tend not to be published, there was no evidence of potential publication bias in the present meta-analysis.This meta-analysis had several strengths. Firstly, the quantitative assessment was based on data from prospective cohort studies. This minimizes the possibility that overall analysis will be influenced due to recall bias, which could be of more common concern in retrospective case–control studies. Also, we had higher statistical power to estimate the relationship between intake ratio of n-3/n-6 PUFA and risk of BC by analyzing summary data from 274,135 participants in 11 prospective studies. Subsequently, serum PL as a biomarker directly reflecting intake ratio of n-3/n-6 fatty acids has the advantage of providing more objective ratio of n-3/n-6 PUFA P for metaregression = 0.103; P for a permutation test = 0.100).Meta-analyses of observational studies are susceptible to methodological and confounding biases inherent in the original studies. Therefore, there are also several limitations considered in our study. Firstly, although prospective studies are more superior to retrospective case–control studies with regards to elucidating causal relationships, selection bias from study populations might be still unavoidable. Volunteers recruited from occupational exposure populations  might not be an unbiased representative of study population. Subsequently, although the fatty acid profile in serum PL, implying absorption and circulation transportation mechanisms, reflects acute dietary intakes over the past few days, it may be a untypical representation of sensitive biomarkers indicating post-absorptive amounts and change at the target tissue. Moreover, any measurement error and resulting misclassification would most likely lead to an attenuation of the true association. Intake of fatty acids was evaluated using dietary questionnaire tools, which could potentially have measurable errors, due to inaccurate fatty acids database and diet reporting bias. Similarly, fatty acids in serum PL,, especially long chain PUFAs, are probably undergoing change during the long follow-up time and storage time, due to The present meta-analysis has also important medical implications of balanced n-3/n-6 fatty acid intake ratio for BC prevention, clinical diagnosis and treatment. Firstly, although many sensitive tumor biomarkers were used for the clinical diagnosis of BC, several potential limitations were still unavoidable. In addition, the BC inflammatory microenvironment plays a major role in growth, invasiveness and resistance to therapy. However, modulation ratio of n-3/n-6 fatty acids in food, as a potent inducer of metabolic responses, could provide an effective means to alter ratio of n-3/n-6 fatty acid in tumor tissue and thereby possibly affect tumor growth, by sensitizing cancer cells to chemotherapy and increasing resistance of normal cells to the toxic effects. Finally, diet intervention could represent a clinically relevant adjuvant therapy in patients with BC, based on the general tenet of clinical nutrition.In summary, findings from this meta-analysis of prospective studies provide a conclusive evidence to support increase intake ratio of n-3/n-6 PUFAs for BC prevention among study females. Quantitative conclusions from the present study showed that per 1/10 increment of n-3/n-6 ratio in diet, there was a 6% reduction of BC risk among study populations . This important evidence highlights the need to promote nutritional education programs, stressing the need to increase the consumption of food rich in n-3 PUFAs (marine foods), decrease the consumption of food rich in n-6 PUFAs (vegetable oils and processed foods), to ultimately improve intake ratio of n-3/n-6 PUFAs. However, no conclusive evidence from USA populations can be obtained and this needs to be,further confirmed by prospective population-based study on tissue biomarker of n-3/n-6 PUFAs ratio, and larger randomized controlled trials are required to determine whether higher intake ratio of n-3/n-6 PUFAs will have beneficial effects on BC risk, or improve the prognosis of patients with BC.The authors declare that they have no competing interests.BY and XR had full access to all of the data in the study, and take responsibility for the integrity of the data and the accuracy of the data analysis. BY XR and DL participated in study conception and design. YF and JG carried out literature research, study identification, data extraction and organization. BY and RX performed study quality assessment and statistical analysis. BY finished drafting of the manuscript; DL XR and BY carried out critical revision of the manuscript. All authors read and approved the final manuscript submitted for publication and are guarantors for the study.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/14/105/prepubCompleted reporting items for systematic review and meta-Analysis (MOOSE Checklist) of present study. Information as to how this article was conducted.Click here for fileCompleted data extraction from eligible original studies. Information as to how the available data were extracted by using the standard extraction forms.Click here for fileTable S2. Study quality assessment of included prospective studies by Newcastle-Ottawa Scale. Figure S1. Dose–response trend between serum PL ratio of n-3/n-6 PUFAs and breast cancer risk among USA females. Figure S2. Contour-enhanced funnel plots for association of ratio of n-3/n-6 PUFA with BC risk.Details on data synthesis and analysis in this meta-analysis. Studies excluded after scrutiny with reasons for exclusion. Click here for file


P < 0.001), particularly in prostate cancer , while no association was found in breast  and colorectal cancers . When we stratified analysis by ethnicity, significant association was found in Caucasians . Further well-designed with large samples and other various cancers should be performed to validate our results.Numerous studies have investigated association between the germline HOXB13 p.Gly84Glu mutation and cancer risk. However, the results were inconsistent. Herein, we performed this meta-analysis to get a precise conclusion of the associations. A comprehensive literature search was conducted through Medline (mainly Pubmed), Embase, Cochrane Library databases. Crude odds ratios (ORs) and their 95% confidence intervals (CIs) were calculated by STATA 12.1 software to evaluate the association of HOXB13 p.Gly84Glu mutation and cancer susceptibility. Then, 25 studies including 51,390 cases and 93,867 controls were included, and there was significant association between HOXB13 p.Gly84Glu mutation and overall cancer risk (OR = 2.872, 95% CI = 2.121–3.888,  Cancer is a serious problem endangering the human health and lives. Based on the reports from International Agency for Research on Cancer (IARC), cancer has become the second leading cause of mortality in developing countries, which has exceeded the mortality caused by cardiovascular incidences and become the leading cause of mortality in developed countries , 2. TotaHOXB13, which encodes the transcription factor 13, belongs to the HOXB gene cluster at chromosome 17 , involveFigure 2I = 62.8%, Pheterogeneity < 0.0001). Considering that, random-effects model was used to examine the association between HOXB13 p.Gly84Glu mutation and overall cancer susceptibility  and then random-effect models was used to verify the relationship between HOXB13 p.Gly84Glu mutation and prostate cancer risk. The results presented that HOXB13 p.Gly84Glu mutation contributed to the susceptibility of prostate cancer  were found when we analyzed the association between HOXB13 p.Gly84Glu mutation and the risk of the other two kinds of cancers mentioned above. HOXB13 p.Gly84Glu mutation was not contributed to the development of breast cancer  .A comprehensive computerized literature search was conducted through Medline (main Pubmed), Embase, Cochrane Library, Web of Science, Wanfang and China National Knowledge Infrastructure (CNKI) databases for related research studies reported the association between HOXB13 p.Gly84Glu mutation and cancer susceptibility. Furthermore, we also searched related studies manually from the references of reviews and articles reported the same topics. Combinations of searching terms were used as follows: “HOXB13 p.Gly84Glu”, “HOXB13 rs138213197”, “single nucleotide polymorphism, SNP or variation, mutation” and “cancer or carcinoma or tumor or neoplasms”. In order to get a precise conclusion, no sample size and language limitations were set, hoping that we can identify all the studies that examined the association of HOXB13 p.Gly84Glu mutation and cancer risk.The following criteria should be met of each included studies. (1) reported the association between HOXB13 p.Gly84Glu mutation and cancer risk; (2) used case-control design; (3) all the patients in cases group should be diagnosed by histochemical results or other gold diagnostic standers; (4) provided sufficient data of HOXB13 p.Gly84Glu mutation carriers or non-carriers, or other information such as ORs with 95% CIs for statistical analysis; (5) if there were several publications with overlapping data, only the latest one with the largest sample size was finally included in this present work. At the same time, if each of the searched studies was in accordance with the following criteria, it must be excluded. (1) human being studies; (2) review, meeting or other types of abstracts, comment, correspondence, letters or letters to the editor; case reports, or case-only studies; (3) not provided the sufficient data to extract.Two independent investigators extracted the essential information according to the selection criteria mentioned above. The key data was listed as follows: the first author's surname, year of publication, country of origin, ethnicity, cancer type, control source [population based (PB) or hospital based (HB)], the total number of cases and controls, genotyping methods, the number of HOXB13 p.Gly84Glu mutation carriers and non-carriers. Any disagreement was resolved by discussion, if not, other authors will join the discussion until consensus was reached.2I test, with the 2I value ranged from 0 to 100%: 2I = 0–25%: no heterogeneity; 2I = 25–50%: moderate heterogeneity; 2I = 50–75%: large heterogeneity; 2I = 75–100%: extreme heterogeneity) [P < 0.10 represents a significant heterogeneity was existed). Both fixed-effects (Mantel-Haenszel) and random-effect (Der Simonian and Laird) models were used to analyze the pooled ORs. The fixed-effects model was used when there was no heterogeneity; otherwise, the random-effects model will be used [P value of less than 0.05 for any test or model was considered to be statistically significant.Crude odds ratio (OR) with corresponding 95% confidence interval (95% CI) were used to calculated to assess the strength of the association between the HOXB13 p.Gly84Glu mutation and cancer risk. Heterogeneity was evaluated by geneity) , 37, and be used , 39. Sub