Cyclin D1 (CCND1) plays a key role in cell cycle regulation. It is a well-established human oncogene which is frequently amplified or overexpressed in cancers. The association between CCND1 G870A polymorphism and cancer risk has been widely assessed. However, a definitive conclusion between CCND1 G870A polymorphism and risk of nasopharyngeal carcinoma (NPC) remains elusive.We firstly performed a hospital-based case-control study involving 165 NPC cases and 191 cancer-free controls in central-south China, and then conducted a meta-analysis with six case-control studies to evaluate the association between NPC risk and CCND1 G870A polymorphism.The case-control study found a significant association between CCND1 G870A polymorphism and NPC risk in various comparison models . Further meta-analysis showed that there was no significant association between CCND1 G870A polymorphism and NPC risk in overall analysis. In the stratified analysis by race, however, significant associations were only found in Caucasians .A significant association between CCND1 G870A polymorphism and NPC risk was found in the central-southern Chinese population. The meta-analysis indicated that CCND1 G870A polymorphism may contribute to the development of NPC in Caucasians. Nasopharyngeal carcinoma (NPC) is considered as one of the rarer cancer forms globally and has a significantly different ethnic and geographic distribution 1-S phase progression of the cell cycle is one of the hallmarks of cancer. The checkpoint of G1/S is frequently altered in many epithelial tumors and may confer growth advantage and increased tumorigenesis 1 transition to S phase is regulated by cyclins, cyclin-dependent kinases and their inhibitors 1/S checkpoint of the cell cycle. CCND1 is a well-established human oncogene. It is frequently amplified or overexpressed in cancers, such as breast cancer, lung cancer, colorectal cancer, melanoma and oral squamous cell carcinomas, through copy number alterations, mutation, or as a consequence of the deregulation of mitogenic signalling downstream of oncogenes such as ERBB2 The abnormality of GOwing to the important of cyclin D1 in the cell division, CCND1 single nucleotide polymorphism (SNP) has been investigated in a variety of tumors, such as uterus, breast, esophagus, lung, colon and head and neck Published studies have also evaluated the association between CCND1 G870A polymorphism and NPC susceptibility This hospital-based case-control study included 165 nasopharyngeal carcinoma (NPC) cases and 191 cancer-free individuals consecutively recruiting from Hunan Tumor Hospital between April 2011 and October 2011. The patients were diagnosed via histopathological evidence and received no treatment before the blood drawing. There were no age, sex, and stage restriction for cases. The selection criteria for controls included no family history of nasopharyngeal carcinoma and frequency matched to cases on age and sex. All recruited subjects were unrelated ethnic Chinese adults, who were resident in Changsha City  or the surrounding regions. Written informed consent was obtained from all participants for the use of their blood samples in the present study. This study was approved by the institutional review board of Hunan Tumor Hospital.ScrF1 at 37°C for 4 hours and visualized by electrophoresis on 3% agarose gels containing 0.5 µg/mL ethidium bromide. The PCR product of 167 base pairs (bp) (the AA genotype) was digested into fragments of 145 and 22 bp for GG and into fragments of 167, 145 and 22 bp for AG. PCR products were randomly selected for DNA sequencing validation.Genomic DNA was extracted from peripheral blood samples with standard procedures  and DNA samples were frozen at −80°C. Genotyping of CCND1 G870A polymorphism was carried out using the polymerase chain reaction-restriction fragment length polymorphism (PCR-RFLP) method. The PCR mixture consisted of 5 pmol of each primer, 1X GoTaq Master Mix  and 2.0 µl extracted DNA at a total volume of 15 µl. The PCR procedure included an initial melting step at 95°C for 1 min, then 35 cycles at 55°C for 1 min, at 72°C for 1 min, and a final extension step at 72°C for 10 min. The primers for PCR amplification were as follows: forward, 5′-GTG AAG TTC ATT TCC AAT CCG C-3’ and reverse, 5′-GGG ACA TCA CCC TCA CCC TCA CTT AC-3’. PCR products (15 µL) were digested with 1 U p<0.05.The Hardy-Weinberg equilibrium was used to test the genotype frequencies of CCND1 G870A polymorphism among the control groups. Chi-square test (χ2) was used to measure the difference in the distribution of genotype and allele frequencies between patients and controls. Odds ratios (ORs) with 95% confidence intervals (95% CIs) for the risk of NPC were calculated to estimate the relative risk using the multivariate logistic regression analysis adjusted by age and gender. Statistical analyses were performed with SPSS 19.0 software. All statistical analyses were two-sided and differences were considered to be statistically significant when Literature searches of the PubMed, Embase, Chinese National Knowledge Infrastructure (CNKI), China Biological Medicine Database and Wanfang Database (up to April 2014) were performed to identify eligible studies. The publication language was restricted to English and Chinese and the combination terms were “CCND1”, “cyclin D1”, “polymorphism”, “mutation or variant” and “nasopharyngeal.” The reference lists of identified studies were manually checked to include other potentially eligible trials. All eligible studies in this meta-analysis met the following criteria: case-control studies in design; investigating the association between CCND1 G870A polymorphism and NPC risk; with genotype distribution data to calculate combined ORs and 95% CIs. The major exclusion criteria were: no control population; abstract, comment, and review; duplicate of earlier publication and no usable genotype frequency data. The following data was extracted from eligible studies: first author, year of publication, country and ethnicity of study population, numbers of cases and controls, and genotype frequency of cases and controls.2I test were used to assess the heterogeneity between studies. 2I<25% indicated low heterogeneity, 25%≤2I≤50% indicated moderate heterogeneity, and 2I>50% indicated large heterogeneity. If QP<0.10 or 2I>50%, the random-effects model (the DerSimonian and Laird method) was used to calculate the pooled OR. Otherwise, the fixed-effects model (Mantel-Haenszel) was selected. The significance of the pooled ORs was assessed via Z-test. The allele model (A vs. G), the co-dominant model (AA vs. GG), the dominant model (AA/AG vs. GG) and the recessive model (AA vs. AG/GG) was performed respectively. Subgroup analysis by ethnicity was also performed. Begg’s funnel plot and Egger’ linear regression test were used to assess publication bias. Hardy-Weinberg equilibrium (HWE) was test in the control groups. In this meta-analysis, all statistical analyses were performed using the software Review Manager (version 5.0) and STATA software (version 12.0).The pooled odds ratio (OR) with 95% confidence interval (CI) was used to evaluate CCND1 G870A polymorphisms and NPC risk. The Q-test and . There were no significant differences between the groups in their gender and age.A total of 356 subjects were included in this case-control study, including 165 patients with NPC and 191 healthy controls. The baseline characteristics were listed in The genotype and allele frequency distributions for G870A among the cases and controls and their associations with risk for cervical cancer are shown in  The characteristics of all the enrolled studies that were performed in the meta-analysis were listed in In total, six studies (844 cases and 1164 controls) published from 2002 to 2014, including our study, were identified to be eligible studies. The detailed process of selecting and excluding articles is presented in The Begg’s funnel plot and Egger’s test were used to estimate the publication bias of the meta-analysis. As the shape of the funnel plots of the all genetic models seem symmetrical, this indicated low publication bias in this meta-analysis Fig. 3.1/S checkpoint of the cell cycle Nasopharyngeal carcinoma is an epithelial malignancy with an unusual ethnic and geographic disparity In recent years, the association between CCND1 G870A polymorphism and NPC risk was also investigated in several studies. The association between CCND1 polymorphisms and NPC risk was firstly reported by Deng et al., which maintained the GG and AG genotypes in NPC patients were significantly higher than those in normal controls There was a published meta-analysis on CCND1 G870A polymorphism and NPC risk 2I values were less than 50% and QP were greater than 0.1 in the Caucasians. However, heterogeneity was still present in the Asians group. The potential factors that may account for the heterogeneity included study design and sample population. Among the four study in the Asians group, the study by Jun Sui et al, Liangchun Shih et al, and our study were designed as a hospital-based case-control study and recruited aged-match controls; whereas the study by Deng Lin et al was not a stringent hospital-based study. The cases of the study by Deng Lin et al were recruited from three hospitals in a city and the controls with lower age were recruited from a university in the same city. The histological types were different in the four studies. All cases of the study by Jun Sui et al were diagnosed as squamous cell carcinomas; most cases (82/84) of the study by Deng Lin et al were poorly differentiated squamous cell carcinomas; whereas histological types of the cases in the study by Jun Sui et al and our study were without restriction. Nevertheless, we are unable to determine the precise factors that account for the heterogeneity due to the limited data.In the present meta-analysis, significant heterogeneity was observed for the association between CCND1 G870A polymorphism and NPC risk among the six studies. By performing subgroup analysis, we found the the There are some limitations in this study. Firstly, the sample sizes of cases and controls were relatively small, especially in the subgroup of Caucasians. There were only two studies on Caucasians and it might result in the false positive findings in our meta-analysis. All results should be interpreted with very caution. Secondly, due to the limited data, we did not carry out subgroup analysis to other factors, which may participate in the progression of disease, such as gene-gene, smoking and other lifestyle. Thirdly, obvious heterogeneity was observed in our meta-analysis. The populations, ethnicity, habits, geographical location and study designs may contribute to the heterogeneity.In conclusion, our case-control study suggested that there was a significant association between the CCND1 G870A polymorphism and NPC risk in the Chinese people. Furthermore, our meta-analysis showed that CCND1 G870A polymorphism was at significantly greater risk for NPC in the Caucasian population. However, these findings should be validated by more studies with large sample sizes, gene-gene, gene-environment interactions, well-designs and more diverse ethnic groups’ data.Figure S1Flow chart of study selection by using electronic database.(PPT)Click here for additional data file.Checklist S1PRISMA 2009 Checklist.(DOC)Click here for additional data file.


MDR1 and GSTs genes might explain individual differences in susceptibility to IBD. We carried out a case-control study to examine the association of MDR1 (C1236T and C3435T), GSTT1, and GSTM1 polymorphisms with the risk of IBD. Subjects were genotyped using PCR-RFLP for MDR1 gene and multiplex PCR for GSTT1 and GSTM1. Meta-analysis was performed to test the association of variant allele carriage with IBD risk. We report that GSTT1 null genotype is significantly associated with the risk of CD  and UC  and can influence Crohn's disease behavior. The interaction between GSTT1 and GSTM1 genes showed that the combined null genotypes were associated with the risk of UC . Furthermore, when compared to combined 1236CC/CT genotypes, the 1236TT genotype of MDR1 gene was associated with the risk of UC . Meta-analysis demonstrated significantly higher frequencies of 3435T carriage in IBD patients. Our results show that GSTT1 null and MDR1 polymorphisms could play a role in susceptibility to IBD.Inflammatory bowel diseases (IBD) are multifactorial disorders resulting from environmental and genetic factors. Polymorphisms in MDR1) and glutathione S-transferases (GSTs) in protecting cells against toxins, xenobiotics, or their metabolites [ MDR1 gene encodes a member of the ABC transporter subfamily B, a transmembrane P-glycoprotein (P-gp) of 170 kDa, which functions as an adenosine triphosphate-dependent efflux transporter pump [ MDR1 gene, the most widely investigated in IBD association studies as well as in other diseases are the 1236C>T , 2677G>T/A , and 3435C>T  with conflicting results in different populations around the world [Inflammatory bowel disease (IBD) is a multifactorial disorder of the gastrointestinal tract including Crohn's disease (CD) and ulcerative colitis (UC). Although considerable progress has been made in the field of IBD research, the underlying etiopathogenesis is still under investigation . It is aabolites . The MDRter pump . P-gp ister pump . The higter pump . Among t GSTM1 (GST-mu 1) and GSTT1 (GST-theta 1) located on chromosomes 1p13.3 and 22q11.2, respectively, are two members of the GSTs family being most frequently studied [ GSTM1 and GSTT1 genes are associated with either decreased or impaired enzyme function [ GSTM1 and GSTT1 genes with the risk of various cancers including bladder, gastric, and oral cancers and chronic myeloid leukemia [ GSTM1 and GSTT1 polymorphisms and the susceptibility of inflammatory and autoimmune diseases such as IBD [ MDR1 and GSTs genes and the risk of IBD has not been examined so far in the Moroccan population. Therefore, our study investigated the role of GSTM1, GSTT1, MDR1 C1236T, and C3435T SNPs in determining disease susceptibility in Moroccan patients.GSTs are phase II xenobiotic metabolizing enzymes. They play a critical role in cellular protection against reactive electrophiles and fatty acid hydroperoxides generated by oxidative stress through the conjugation with reduced glutathione. Therefore, GSTs facilitate the detoxification of cells by limiting tissue damage from free radical attack , 13. GST studied , 15. Comfunction . Severalleukemia –20. Howeh as IBD –23. To thttp://dx.doi.org/10.1155/2015/248060). The local ethics committee approved the study and a written informed consent was obtained from all participants.A total of 110 patients diagnosed with IBD at the Department of Gastroenterology, CHU Ibn Rochd , were selected. Blood samples from 100 blood donors were used as controls. The diagnosis of CD or UC was established according to conventional clinical, endoscopic, radiological, and histological criteria as previously described , 25. Pat GSTM1 and GSTT1, a multiplex polymerase chain reaction (PCR) was performed, in which BCL2 gene was used as an internal control. PCR amplification condition and products sizes were previously described [Genomic DNA was extracted from whole blood using the salting-out method. DNA concentration and quality were analyzed using a NanoDrop 1000 spectrophotometer . Genotyping of C1236T and C3435T SNPs was done by polymerase chain reaction, restrictive fragment length polymorphism. The primer sequences, enzymatic restriction conditions, and digestion product sizes were previously described , 27. To escribed . MDR1 C3435T, C1236T, GSTM1, and GSTT1 polymorphisms and IBD, and (c) the study reporting sufficient data to calculate allele frequencies and odds ratios of cases and controls for carriage of MDR1 3435T and 1236T alleles. Major exclusion criteria were (a) case-only study and review articles and (b) studies without raw data of the MDR1 and GST genotypes.Genetic association studies were included in our meta-analysis if they met the following criteria: (a) a case-control study design, (b) studies that evaluated the association between the MDR1 polymorphisms in IBD patients were identified through the literature search. Eighteen studies that met the inclusion criteria were retrieved in MDR1 C3435T meta-analysis  test and I2 statistics was used to quantify the between-study heterogeneity effect. When a significant Q test  indicated heterogeneity across studies, data were recombined using a random-effects model to estimate common ORs. The meta-analyses were conducted by MedCalc v.11.6.1.0 software.Twenty-seven case-control studies investigatinganalysis . Five ofanalysis . GSTM1 acontrols . The risP value less than 0.05 was considered as statistically significant.Data analysis has been carried out using the statistical package SPSS version 16 . Differences in distribution of demographic and clinical features of patients with respect to their genotypes were done by chi-square test or Fisher Exact Test. The same tests were used to compare the distribution of genotypes between patients and controls as well as assess the Hardy-Weinberg Equilibrium in MDR1 gene. Odds ratio (OR) with a confidence interval (CI) of 95% was calculated to measure the strength of association between C1236T, MDR1 C3435T, GSTT1, and GSTM1 and the risk of inflammatory bowel disease. A P > 0.05). This finding remained valid for the allele frequencies (P > 0.05). In P = 0.03). However, the dominant model showed no particular effect whatever the type of considered SNP is. As presented in  GSTM1 null genotype frequency was higher in CD patients without being statistically significant . The same trend was observed in UC patients . Surprisingly, we noticed that the GSTT1 null genotype was significantly associated with the risk of CD  and the risk of UC . Furthermore, the interaction between GSTM1 and GSTT1 showed that the combined null genotype  was associated with the risk of UC at the limit of the statistical level  , and C3435T (exon 26) in MDR1 gene showed no particular trend of association (data not shown). On the other hand, frequency of the stricturing form was statistically higher in CD patients carrying the GSTT1 null genotype compared to the penetrating or inflammatory forms . However, the association between smoking status  and CD/UC was not influenced by genetic polymorphisms in genes encoding the GSTs metabolizing enzymes  . The dis enzymes . MDR1 C3435T combined to our results, we observed a significant association between the T allele and IBD risk .Based on the studies published onIBD risk . Meta-anIBD risk . On the nd GSTT1 , a signi MDR1 C1236T, MDR1 C3435T, GSTM1, and GSTT1 polymorphisms on the risk of CD and UC disease. No genotype-phenotype correlation was observed between clinical characteristics of patients and the different genotypes of C1236T and C3435T polymorphisms in MDR1 (data not shown). Consistent with our results, Fischer et al. reported a lack of association between C3435T and IBD phenotype in Hungarian patients [ GSTT1 was higher in CD patients with the stricturing form. In contrast to our findings, Karban et al. reported that neither GSTM1 null nor the GSTT1 null genotypes were found to be associated with CD or UC phenotypes [ MDR1 C1236T and MDR1 C3435T in our patients and controls were statically comparable (P > 0.05). Similar findings for the C3435T SNP were reported by Wang et al., in a meta-analysis based essentially on Caucasians [ MDR1 C1236T was statistically associated with the risk of UC (P < 0.03). The discrepancy between our findings and those studies might be explained not only by the difference in minor allele distribution of MDR1 C1236T and MDR1 C3435T in our population but also by the relative small sample size of the present study. We have previously reported the existence of linkage disequilibrium between C1236T and C3435T in MDR1 gene in our population [ MDR1 polymorphisms. On the other hand, we have interestingly noted that the GSTT1 null genotype was found to be significantly associated with the risk of both CD  and UC disease . Consistent with our finding, Mittal et al. have reported a significant association of GSTT1 null genotype with susceptibility to CD and UC in Indian population; however, they also found an association of GSTM1 null genotype with the risk of UC that we could not replicate [ GSTT1 and GSTM1 were associated with the risk of CD and UC in Israeli population [ GSTM1 null/GSTT1 null genotype was associated with the risk of UC at the limit of statistical level. Both previously reported studies have shown the association of the double deletion in GSTM1 and GSTT1 genes with IBD [ GSTs polymorphisms and IBD has been rarely discussed; more association studies are needed to validate the conclusions.In the present case-control study, we investigated the potential influence ofpatients . In contpatients . We noteenotypes . The freucasians . Brinar ucasians . Furtherucasians . It is npulation . It is wpulation . The intwith IBD , 23. A c MDR1, GSTM1, and GSTT1 polymorphisms with the risk of IBD in a sample of the Moroccan population. It follows from the present case-control study that GSTT1 null genotype and MDR1 C1236T in the recessive model are associated with the risk of IBD. Crohn's disease behavior was influenced by the GSTT1 null genotype. Moreover, the combined null genotype of GSTT1 and GSTM1 was associated with the risk of UC with a limited effect.This is the first study to examine the association ofGenotypic distribution of MDR1 polymorphisms according to Crohn's disease and Ulcerative colitis patient's demographic and clinical characteristics are presented in Supplementary Table 1 and 2 respectively. In addition, genotypic distributions of GSTM1 and GSTT1 with respect to demographic and clinical characteristics of Crohn's disease and Ulcerative colitis patients are presented in Supplementary Table 3 and 4 respectively.


The aim of this meta-analysis was to compare the long-term efficacy of diet plus exercise (D + E) vs. diet (D), D + E vs. exercise (E) and D vs. E on anthropometric outcomes and cardiovascular risk factors in overweight and obese participants.2 and a minimum intervention period including follow-up of ≥12 months. Outcomes of interest were as follows: anthropometric parameters, blood lipids, blood pressure and cardiorespiratory fitness. Pooled effects were calculated using pairwise random effects and Bayesian random effects network meta-analysis. Results of the corresponding fixed effects models were compared in sensitivity analyses.Electronic searches were performed in MEDLINE and the Cochrane Central Register of controlled trials. Inclusion criteria were as follows: body mass index ≥25 kg/mOverall, 22 trials (24 reports) met the inclusion criteria and 21  of them were included in the quantitative analysis. As compared with D, D + E resulted in a significantly more pronounced reduction in body weight , and fat mass , waist circumference , and fat mass  was in favour of combined diet and exercise, respectively. Comparing E vs. D, diet resulted in a significantly more pronounced decrease in body weight , and fat mass . D + E yielded also the greatest reductions with respect to blood lipids and blood pressure when compared to single applications of D and E, respectively. Results from the network meta-analyses confirmed these findings.Moderate-quality evidence from the present network meta-analysis suggests that D + E can be highly recommended for long-term obesity management. Furthermore, the evidence suggests a moderate superiority of D over E with respect to anthropometric outcomes.CRD42013003906PROSPERO  Data extraction was performed by one author (LS).Separate pairwise meta-analyses were first used to compare all lifestyle interventions. Network meta-analysis was then used to synthesize all the available evidence . Networkχ2 test. The I2 parameter was used to quantify any heterogeneity: I2 = [(Q - d.f.)]/Q × 100%, where Q is the χ2 statistic and d.f. is its degrees of freedom. A value for I2 > 50% was considered to represent substantial heterogeneity , WC , WHR  and FM  was significantly more pronounced in the D + E group as compared to D, respectively. Furthermore, the D + E group revealed significantly more prominent increases in cardiorespiratory fitness (measured as VO2 max)  and HDL cholesterol  as well as decreases in TG  and DBP . In contrast, changes observed for TC, LDL-C and SBP did not differ significantly between both groups.The weighted mean difference in change of BW , WC , WHR  and FM  could be observed in the D + E group. Rise of VO2 max was significantly more pronounced in the D + E group as well . With respect to blood lipids, TC , LDL-C , DBP  and SBP  were reduced more substantially following combined D + E when compared to single E interventions. No significant differences could be observed for HDL-C and TG.Comparing D + E vs. E, a significantly more distinctive reduction in BW , FM , HDL-C  and SBP  were significantly more pronounced in the D group. WC, WHR, TC, LDL-C and cardiorespiratory fitness were not affected in a different fashion by either D or E.Following D vs. E comparisons, reductions in BW [MD: -2.93 kg (95% CI -4.18 to -1.68), Additional file Both D + E and D were significantly more effective in reducing BW and FM when compared to E alone. D + E turned out to be the most effective lifestyle intervention with respect to reduction of BW, WC, FM, TC, LDL-C, TG, DBP, SBP and increasing HDL-C. D + E resulted in a high (>75%) probability to be the best for most outcomes. There is greater uncertainty regarding which treatment is the best for HDL-C, although again D + E yielded the highest probability of being the best. D turned out be the second effective lifestyle intervention for BW, WC, FM, TC and DBP (>75% probability).p value = 0.02) although this might be due to chance since several p values for inconsistency are being calculated  between D + E  vs. D , D + E (79/714) vs. D (82/715) (p = 0.96) and D + E (87/701) vs. D (84/706) (p = 0.34). Eight trials (nine reports) reported random sequence generation [The dropout rates ranged from 0% to 65%, with 11 out of 22 trials reporting dropout rates <10% (Table 6) p = 0.. Eight tneration ,47,50-52neration ,52. Noneneration ,47,50-526) p = 0.. Eight tneration ,46,50-522 max. Including only participants ≥50 years of age  were performed for obese, study length ≥24 months, older participants (age: ≥50 years) and low risk of bias studies. Comparison of long-term vs. short-term trials resulted in smaller reductions in body weight , compared to the random effects model. Furthermore, the fixed effects model showed a significant reduction in WHR, comparing D + E vs. D/E, and increase in HDL-C compared to E. Again, we note that the fixed effects model was not always considered suitable . Funnel plots were generated only if specific outcome measures were provided by at least ten different trials  . The plo2 max.The primary findings of this systematic review and meta-analysis of long-term trials are that a combination of D + E is more effective in reducing BW compared to either D or E in overweight and obese participants. Furthermore, improvements in WC, WHR and FM were significantly more pronounced following D + E compared to either D or E. The combination of D + E is also the most powerful lifestyle intervention to reduce blood lipids and blood pressure. Compared to E, a dietary intervention resulted in significantly greater reductions in body weight, fat mass and systolic blood pressure. Pooling both direct and indirect evidence on D + E, D and E via network meta-analysis demonstrated that D + E was the most efficacious exercise intervention regarding its impact on BW, WC, FM, TC, LDL-C, TG, DBP and VOThe findings of this meta-analysis, i.e. diet being significantly more effective in reducing BW than exercise, are supported by results of a 2006 Cochrane review  and are Previous studies reported that a 5% reduction of BW is associated with a reduced risk of T2D incidence and other metabolic disorders. A 5-kg weight loss over time could account for a 55% reduction in the risk of diabetes over the mean of 3.2 years of follow-up in a high-risk population . RegardiWith respect to blood lipids, a meta-analysis of 70 studies indicate that each kilogram of weight loss was associated with a 1.9 mg/dl decrease in TC and a 0.77 mg/dl decrease in LDL-C, respectively . FurtherThe predominant dietary intervention implemented in the included trials was either an energy-restricted low-fat diet or an energy-balanced moderate-fat diet. In general, the composition of diets was approximately at least 500 kcal below the estimated energy need, and fat intake was ≤30% of total energy content. In the D + E and D trials, 1,200 kcal for women and 1,500 kcal for men were generally prescribed. In the dietary intervention trials, general guidelines for physical activity were recommended. However, in the D + E trials, specific goals for physical activity/exercise were implemented. Network meta-analyses provides evidence that a combination of aerobic and resistance training should be recommended in the prevention and treatment of overweight, obesity and associated diseases ,10.Cardiorespiratory fitness is associated with cardiovascular mortality and cancer in men and women ,64. A poThe present systematic review has several limitations. A major limitation is that no search for unpublished studies and any additional data sources and strategies  was performed. Another limitation is the fact that not all potential effect modifiers were accounted for. Often participants in different arms of trials did not receive equal numbers of contacts. It could be argued that contact time rather than the specific elements of the intervention affected participant’s weight and cardiovascular risk factor outcomes. Heterogeneity could be observed for some outcome parameters, probably introduced by differences between trials, including different D + E regimens. Publication bias cannot be excluded to affect the results on any meta-analysis; however, formal statistical testing did not suggest publication bias for the current analysis. Had there been evidence of publication bias and if more studies were available, regression techniques could be used to adjust for this .Although many studies included in our analysis had a substantial dropout rate  in combination with dietary interventions for the prevention and treatment of overweight and obesity.The authors declare that they have no competing interests.LS extracted the data. LS, SD and GH conducted the data analysis, interpreted the results and drafted and finalized the manuscript. All authors read and approved the final manuscript.Full search strategy: PUBMED. Table S1: sensitivity analysis (low risk of bias). Table S2: sensitivity analysis (study length: ≥24 months). Table S3: sensitivity analysis . Table S4: sensitivity analysis (obese participants). Table S5: sensitivity analysis (participants ≥50 years of age). Table S6: sensitivity analyses . Table S7: model fit statistics for the random and fixed effects network meta-analysis models. Table S8: estimates of the effects of diet + exercise vs. exercise from direct and indirect evidence in node split models. Figure S1: network graph. Figure S2: funnel plot for body weight (diet + exercise vs. diet). Figure S3: funnel plot for body weight (diet + exercise vs. exercise). Figure S4: funnel plot for body weight (diet vs. exercise).Click here for file


RRs) and restricted cubic spline model was adopted for the does-response analysis.Several epidemiological studies have investigated the association between breastfeeding and endometrial cancer (EC). However, the results of the studies are controversial. Thus, we conduct this meta-analysis to explore the association between breastfeeding and EC and to evaluate the possible does-response relationship between duration of breastfeeding and EC. PubMed, Web of Science, Chinese National Knowledge Infrastructure, China biology medical literature database, Wan fang databases and Database of Chinese Scientific and Technical Periodicals were searched for eligible observational studies up to 11 July 2015. Random effects model was used to calculate the pooled relative risks , 0.58–0.95). In subgroup analyses, a significant association of breastfeeding with EC risk was found in Asia , and an inverse association of breastfeeding with EC risk was found in cohort studies . The results were also significant after adjusted for hormone use  and body mass index . A linear relationship was found of breastfeeding with EC (p for nonlinearity = 0.93), and it indicated that EC risk decreased by 1.2% for one month increment of breastfeeding. This meta-analysis indicates that long term breastfeeding might be associated with decreased risk of EC.Fifteen articles with 623570 participants were identified. The  Endometrial cancer (EC) is one of the most common malignancies among women. There were 320,000 new cases diagnosed in 2012 worldwide . PresentBreastfeeding is an essential biological function of humans and the beneficial effects of breastfeeding for both mother and child are widely acknowledged –14. MetaSeveral epidemiological studies have investigated the association between breastfeeding and EC –22. HoweRRs concerning the relation between breastfeeding and EC. Eventually 15 published articles [We identified 9109 articles by the search strategy and 3 articles by searching reference lists, of which 9000 articles were excluded after review of the title/abstract or which were duplicated  was 0.74 (95% CI 0.58–0.95). We have separated the results by ever breastfeeding and the duration of breastfeeding. The pooled RR of 14 studies [CI 0.72–1.06) for the ever breastfeeding compared with the reference group. The association between breastfeeding and risk of EC is provided in Figure The pooled  studies , 26–32 wRR = 0.57, 95% CI 0.37–0.87), but the association was not significant in Europe  and in North-America .We found a strong and significant association of breastfeeding with EC risk in Asia , and the pooled RR for case-control studies was 0.78 .Inverse association was found for cohort studies. The pooled We found no significant association in stratified analysis by adjustment (yes or no) for the following covariates: postmenopausal hormone replacement therapy use, oral contraceptive use, body mass index (BMI), menarche age and menopausal status. Moderate or lower between-study heterogeneity was found in all analyses. Table p for nonlinearity = 0.93). Compared with 0 month, the RRs (95% CI) of EC were 0.99 (0.99–1.00), 0.92 (0.86–0.98), 0.83(0.74–0.94) and 0.79 (0.69–0.90) for 0.5, 6, 14 and 18 months separately. The dose–response analysis suggested that EC risk decreased by 1.2% for one month increment of breastfeeding . The ‘leave one-out’ sensitive analysis [2I = 42.1%, p > 0.05) was found, and the pooled RR was 0.78 (95% CI 0.64–0.96).Moderate between-study heterogeneity was found in this analysis. Exploratory univariate meta-regression was performed with the covariates of published year, study area, study type, cases and sample size. However, results showed that no covariate had a significant impact on between-study heterogeneity .p = 0.097)  of EC did not substantially change in the sensitivity analysis. Subgroup analysis by study design showed the inverse association between breastfeeding and EC for cohort studies. These results proved that our results were stable and credible.To our knowledge, this is the first meta-analysis to examine the association between breastfeeding and EC. From the results we found that breastfeeding was associated with a statistically significant decreased risk of EC, and there was a notable linear relationship between duration of breastfeeding and risk of EC. Dose–response analysis revealed that EC risk decreased by 1.2% for 1 month increment of breastfeeding. The summary RR=0.57, 95% CI 0.37–0.87). In Asia, Europe and North America, the traditional culture and people's lifestyles are different. First, more Asians tend to choose breastfeeding as the way to feed children with respect to the Europeans and North Americans, and the duration of breastfeeding is longer [We found a strong and significant association of breastfeeding with EC risk in Asia  was found. The studies that we excluded in the ‘leave one-out’ sensitive analysis contained a relatively small number of cases, and the adjustment factors in the study of Rosenblatt KA et al [RR was 0.78 (95% CI 0.64–0.96) after the ‘leave one-out’ analysis.Between-study heterogeneity is common in meta-analysis , and it KA et al  was lessRR of cohort studies was 0.62 (95% CI 0.41–0.94).A major strength of this meta-analysis is the large number of included participants, allowing us to get the stable results. Second, most included studies had adjusted for potential confounders, increasing the credibility of the results. Third, dose–response analysis was performed to better describe the association of breastfeeding time with EC risk. In addition, the subgroup analysis by study design strongly identified the effect of breastfeeding on EC, the pooled However, the limitations of our study should also be considered. First, further adjustments need to be conducted to clarify the independent role of breastfeeding to EC. Second, the adjustment factors of each study were different. Third, majority studies of this analysis were case-control design. Considering the recall bias of case-control studies, the effect of breastfeeding on EC requires confirmation in large cohort studies.In summary, this meta-analysis confirms an inverse association between breastfeeding and EC risk especially in Asia. Our findings also confirm that longer time of breastfeeding may reduce the risk of EC. The association of breastfeeding and EC deserves further investigation.The initially relevant studies from PubMed, Web of Science, Chinese National Knowledge Infrastructure, China biology medical literature database, Wan fang databases and Database of Chinese Scientific and Technical Periodicals were identified up to 11 July 2015. All the articles were restricted to English or Chinese language. The following strategy was used in this search: ((((((reproductive) OR reproduction))) OR ((breastfeeding) OR lactation))) AND ). In addition, the reference lists from retrieved articles were also reviewed to identify relevant studies , 46.RR) or hazard ratio with 95% confidence interval (CI) were available. (3) The number of cases and participants or person-years for different duration of breastfeeding must be provided in order to do the dose–response analysis. If multiple articles were published from the same study, we choose the complete one or the study which can provide sufficient detail of data.All studies were reviewed independently by the first and the second authors. If the two authors disagreed about the eligibility of an article, it was resolved by consensus with the third author. The studies must meet the following inclusion criteria: (1) case-control or cohort studies published as original studies to evaluate the association between breastfeeding and risk of EC incidence. (2) Odds ratio (OR), relative risk . We extracted RRs adjusted with the most confounders. For dose–response analysis, the number of cases, participants (person-years), and RR (95% CI) for each category of breastfeeding were extracted, as well as the different duration of breastfeeding [The following items were extracted from each study: the first author's name, sample size, EC cases, study type, publication year, continent, follow-up years for cohort studies, adjustment for potential confounding, tfeeding . The stuRR with 95% CI by the inverse variance-weighted method. We used the Q test and the 2I statistic to assess the heterogeneity among studies [Pooled measure was calculated as the inverse variance-weighted method of the logarithm of  studies . The REM studies .Meta regression and subgroup analyses were conducted to explore the possible sources of between-study heterogeneity. The restricted maximum likelihood method (REML) was used to estimate the additive (between-study) component of variance tau^2 in meta-regression. The ‘leave one-out’ sensitive analysis  was carrRR estimates across duration of breastfeeding. First, we created a restricted cubic spline model with three knots at the 10 th, 50 th, and 90 th centiles. Then the REML method was used to combine the study-specific estimates in a multivariate random-effects meta-analysis [p value for nonlinearity was calculated. All statistical analyses were conducted by Stata V.12.0 . A 2-sided and a p < 0.05 was considered statistically significant.For the does-response analysis, the method described by Orsini et al  was usedanalysis . By test


Pilot studies have estimated cancer incidence in patients with systemic lupus erythematous (SLE). However, the results have been inconclusive. To ascertain the correlation between SLE and malignancy more comprehensively and precisely, we conducted a meta-analysis.2 index. Publication bias was assessed by Egger’s test.PubMed, the Cochrane Library and Embase databases through June 2014, were searched to identify observational studies evaluating the association between SLE and malignancy. The outcomes from these studies were measured as relative risks (RRs). A random or fixed effects model was chosen to calculate the pooled RR according to heterogeneity test. Between-study heterogeneity was assessed by estimating I2, 57.6% vs 74.3% vs 67.7% vs 82.3%). No apparent publication bias was detected except for NHL studies.A total of 16 papers, including 59,662 SLE patients, were suitable for the meta-analysis. Of these papers, 15 reported RRs for overall malignancy, 12 for non-Hodgkin lymphoma (NHL) and lung cancer, 7 for bladder cancer, 6 for Hodgkin lymphoma (HL) and leukemia, 5 for skin melanoma, and liver and thyroid cancers, 4 for multiple myeloma (MM), and esophageal and vaginal/vulvar cancers and 3 for laryngeal and non-melanoma skin cancers. The pooled RRs were 1.28  for overall cancer, 5.40  for NHL, 3.26 for HL, 2.01 for leukemia, 1.45 for MM, 4.19 for laryngeal cancer, 1.59  for lung cancer, 1.86 for esophageal cancer, 3.21 for liver cancer, 3.67 for vaginal/vulvar cancer, 2.11 for bladder cancer, 1.51 for non-melanoma skin cancer, 1.78 for thyroid cancer, and 0.65 for skin melanoma. Only the meta-analyses of overall malignancy, NHL, and liver and bladder cancers produced substantial heterogeneity (IOur data support an association between SLE and malignancy, not only demonstrating an increased risk for NHL, HL, leukemia, and some non-hematologic malignancies, including laryngeal, lung, liver, vaginal/vulvar, and thyroid malignancies, but also a reduced risk for skin melanoma. Although an increased risk of MM, and esophageal, bladder and non-melanoma skin cancers was identified from the accumulated data in these studies, this observation requires confirmation. Systemic lupus erythematous (SLE) is one of the most common systemic autoimmune rheumatic diseases. It often affects multiple organ systems and has a broad range of clinical and laboratory manifestations–3 and ocIntrinsic immune system defects in SLE, combined with exposure to cytotoxic medications, foster the emergence of site-specific cancers –8. In thWe searched PubMed, the Cochrane Library and Embase databases through June 2014 for English articles using the following words , which were retrieved from all eligible articles and the references were reviewed to identify additional relevant studies: [‘autoimmune diseases’ OR ‘systemic lupus erythematous’ OR ‘SLE’] and [‘lymphoma’ OR ‘malignancy’ OR ‘cancer’ OR ‘neoplasm’ OR ‘tumor’] AND [‘case-control’ OR ‘cohort’].Eligible studies fulfilled the following criteria: (1) case-control or cohort study; (2) SLE as one of the exposure interests; 3) cancer as one outcome of interest; (4) general population as the control group; (5) relative risk (RR), standardized incidence rate (SIR), or standardized morbidity rate (SMR) with 95% confidence interval (CI) available . Reviews  cancer a,25–27 orData extraction was conducted by one investigator (L-HC) and checked by two investigators (G-XX and PL). Discrepancies were resolved through team consensus. The collected data included general information , cohort of SLE characteristics , and results . The quality of each study was evaluated independently by the above three authors using the Newcastle-Ottawa scaleS1 TableS1 Table.The preferred method of data presentation for cohort studies are the calculated RRs compared with the general population, generally estimating the age- and gender-adjusted SIRs or SMRs. SIRs or SMRs are estimated as the ratio of the observed over the expected number of cases for an exposed population. The expected number is calculated as the product of gender-, age-, and calendar period-specific person-years of follow-up, and the corresponding incidence of malignancy in the general population. The 95% CI is calculated assuming a Poisson distribution for the observed cases.2 (significance level at P <0.1) [2 estimation, random effects models were used. The Galbraith plot was used to detect potential source of heterogeneity [All analyses were performed with Stata 12.0 software. Fixed (Mantel-Haenszel) or random effects (DerSimonian-Laird) models were used to calculate the pooled effect estimates. Between P <0.1) . In the ogeneity . We alsoogeneity . The preA total of 2,350 articles met the defined search criteria, the titles and/or abstracts of which were screened to identify the potentially relevant articles. Of them, 29 publications were further analyzed. Finally, 16 cohort studies were selected for the meta-analysis . All of 2 = 57.6%). The Galbraith plot showed that three studies might be the major source of the heterogeneity , moderate heterogeneity in the population-based one (Ι2 = 45.4%). The studies adopting the American College of Rheumatology criteria for the classification of SLE [2 = 50.8%). When cancers diagnosed in the first year of SLE were excluded, the summary risk estimate was 1.25, otherwise, it increased . Males were at a higher risk to develop a cancer  than females . Egger’s test was used to evaluate publication bias for the meta-analysis of overall malignancy. There was no apparent asymmetric distribution occurring in it  conferreCR, 1982) did . The possible source of heterogeneity might be attributed to two studies [2 = 78.5%) while the North American studies didn’t (2 = 84.5%). The heterogeneity among the population-based studies (Ι2 = 4.4%) might not be important. In the analysis stratified by diagnostic criteria for SLE, the risk of NHL (pooled RR and 95% CI) was significantly decreased in the subgroup with ACR criteria, 4.97 (3.94–6.27), compared to that with ARA criteria (2 = 2.3%). When NHLs in the first year of SLE were excluded, the summary risk estimate was 3.92 with moderate heterogeneity (Ι2 = 34.4%), otherwise, it increased  with considerable heterogeneity , substantial heterogeneity among the liver cancer ones (Ι2 = 67.7%), considerable heterogeneity among the bladder cancer ones (Ι2 = 82.3%). The Galbraith plot showed that no study contributed to the heterogeneity among the liver cancer studies and one study [As for non-hematologic malignancies, an increased risk was discovered in lupus patients for laryngeal, lung, esophageal, hepatic, vaginal/vulvar, bladder, non-melanoma skin and thyroid malignancies while a decreased risk was uncovered for melanoma skin cancer . Twelve ne study  might bene study  was exclne study  was omitFrom The present study is by far an ever-comprehensive meta-analysis of SLE and cancer risk, providing more precise risk estimates of various malignancies including some rarer cancers such as laryngeal and esophageal cancer in SLE. It demonstrates that NHL, HL, leukemia, and laryngeal, lung, liver, vaginal/vulvar, and thyroid malignancies are more frequently observed in lupus patients than in the general population. In addition, there is a potentially increased risk of MM, and esophageal, bladder and non-melanoma skin cancers. Conversely, there is a reduced risk of melanoma skin cancer. Hence, the overall risk of malignancy in SLE is slightly elevated, compared to the general population.Nevertheless, these findings may be influenced by some confounding factors. The subgroup analyses for overall cancer and NHL have shown that the risk for hospital-based cohorts is higher than population-based ones. A potential persuasive explanation is that hospitalized lupus patients have a more severe condition so that a greater number develop into a cancer. European and North American lupus populations are at almost the same risk for the development of a cancer as Asian populations but at a much lower risk to develop NHL. The SLE patients with ACR criteria confer a lower risk to develop a cancer, especially NHL, than those with ARA criteria, revealing that preliminary criteria probably overestimates the true risk owing to lack of specificity. From thThere have been some hypothesized explanations for the differences in the risks of certain malignancies in lupus patients versus the general population. Significant concern is that the baseline immune system defects in this disease, combined with exposures to cytotoxic and immunosuppressive medications, may increase malignancy susceptibility .According to our study, lymphoma, especially NHL, is a common malignancy in lupus patients. Possible mechanism may be composed of several aspects. First, diffuse large B cell lymphoma, originating from activated lymphocytes, has been shown to be the most common NHL subtype arising in SLE, suggesting that chronic inflammation might heighten lymphoma risk in autoimmune diseases like SLE . Second,With respect to non-hematologic malignancies, lung cancer risk is increased by 59% in lupus patients compared to the general population. Two potential issues might explain the association. Firstly, it is confirmed that there is a case history of more pack-years in lupus patients developing lung cancer than the general population with lung cancer . MoreoveThere are two main limitations which should be considered when interpreting the data from the present study. First, the primary limitation is the heterogeneity especially in the meta-analyses of overall malignancy, NHL, and liver and bladder cancers. Although subgroup analyses have indicated that some confounding factors don’t change the risk trends for overall malignancy and NHL, the other factors like SLE management strategies, especially cytotoxic drugs use, coexistence of overlap syndrome, and lifestyle factors such as smoking have been unavailable from some studies so that these subgroup analyses couldn’t be conducted, which may have had an impact on the estimated accuracy. Second, even though no significant evidence is found by egger’s test in all the meta-analyses with the exception of NHL, we are not able to completely exclude the publication bias because we didn’t unveil unpublished data and only collected the published articles in English.Despite the undeniable limitations, owing to a vast sample size, our study support an association between SLE and malignancy, not only demonstrating an increased risk for NHL, HL, leukemia, and some non-hematologic malignancies including laryngeal, lung, liver, vaginal/vulvar, and thyroid malignancies, but also a reduced risk for skin melanoma. Although an increased risk of MM, and esophageal, bladder and non-melanoma skin cancers was identified from the accumulated data in these studies, this observation requires confirmation. The findings highlight the importance of follow-up and cancer screening in lupus patients. Future studies should focus on the underlying mechanisms related to the development of site-specific cancers, especially MM, and laryngeal, esophageal and melanoma skin cancers, in patients with SLE and warrant correlation with our epidemiologic data.S1 PRISMA Checklist(DOC)Click here for additional data file.S1 Fig(DOC)Click here for additional data file.S2 Fig(DOC)Click here for additional data file.S3 Fig(DOC)Click here for additional data file.S4 Fig(DOC)Click here for additional data file.S5 Fig(DOC)Click here for additional data file.S6 Fig(DOC)Click here for additional data file.S7 Fig(DOC)Click here for additional data file.S8 Fig(DOC)Click here for additional data file.S9 Fig(DOC)Click here for additional data file.S10 Fig(DOC)Click here for additional data file.S11 Fig(DOC)Click here for additional data file.S12 Fig(DOC)Click here for additional data file.S13 Fig(DOC)Click here for additional data file.S14 Fig(DOC)Click here for additional data file.S15 Fig(DOC)Click here for additional data file.S16 Fig(DOC)Click here for additional data file.S17 Fig(DOC)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.


To provide a precise quantification of the association between alcohol and tobacco consumption trends in head and neck cancer patients over the past 45 years.We combined findings from all studies published until March 2014 and evaluated the association between different levels in alcohol and tobacco consumption and head and neck cancers through a meta-analytic approach.We included 28 studies involving 13830 patients with head and neck cancer. In patients with alcohol consumption, the pooled odds ratio (OR) and 95% confidence interval (CI) were 1.29(1.06-1.57), 2.67(2.05-3.48) and 6.63(5.02-8.74) for light drinkers, moderate drinkers and heavy drinkers, respectively. In patients with tobacco consumption, the pooled OR and 95% CI were 2.33(1.84-2.95), 4.97(3.67-6.71) and 6.77(4.81-9.53) for light smokers, moderate smokers and heavy smokers, respectively.The increased alcohol and tobacco consumption trends increased the risk of head and neck cancer over the past 45 years. Tobacco consumption was found to be a stronger risk factor for head and neck cancer than alcohol consumption. Thus, the control should be considered to limit the intake of alcohol and tobacco. The broad ranges of alcohol and tobacco consumption patterns around the world create significant public health and safety problems. Convincing evidence demonstrated that alcohol and tobacco consumption increases the risk of cancer in the breast, colorectum, liver, esophagus and head and neck –3. CigarHead and neck cancers (HNC) including cancers in the oral cavity, pharynx and larynx, is one of the most common cancer in the world. There are approximately 300,000 oral cancers, 142,000 pharynx cancers and 157,000 larynx cancers been diagnosed each year . The theWe carried out a systematic literature search in PubMed and ISI Web of Science for articles published before March 2014. We also reviewed references from reviews, meta-analyses and relevant studies for the sake of completeness. The key words used for the literature search was as follows:  AND (smoke OR tobacco consumption) AND . All titles and abstracts were reviewed by two of the authors.The articles which met the following explicit criteria were included: 1. Case–control or cohort studies; 2. The exposure of interest were alcohol and tobacco consumption; 3. Risk estimates [odds ratio (OR) or risk ratio (RR)] and corresponding 95% confidence intervals (CIs) or sufficient data to calculate them were reported; 4. Never and/or occasional  drinking was taken as the reference category; 5. At least three levels of alcohol and tobacco consumption and reporting the OR or RR and 95% CIs or sufficient information to calculate them for each level.For each included study, we extracted details on publication year, country, study name, cancer site, study design, variables adjusted or matched, risk estimates and 95% CIs. Two reviewers assessed the quality of included articles, and resolved doubts or disagreements. When same studies were published in more than one paper, only the most recent article was included in the analysis.Since different studies used different units to measure alcohol consumption, we converted it into a uniform measurement (grams per day (g/day)) and the equivalencies was formulate as 0.8 g/ml = 28g/ounce = 12.5 g/drink. The risk estimates of alcohol consumption were separated into light, moderate, and heavy drinkers, which were ≤12.5 g/day, 12.6–49.9 g/day, and ≥50 g/day of alcohol consumption, respectively . Due to P values were two-sided.Because all included publications were case-control studies, we used ORs to measure the interest. When available, we used adjusted risk estimates; otherwise the unadjusted risk estimates were used. Chi-square statistics was calculated to evaluate the heterogeneity across the studies. When P<0.10, the random effect model was selected, otherwise, the fixed effect model was performed . The EggThe process of selection of articles for inclusion was summarized in I2> 50%). The dose-response analysis was carried out in I2> 50%). We also carried out a dose-response analysis in For the head and neck cancer is a chronic process, the duration of alcohol and tobacco consumption should be considered for a risk factor. So we showed results of pooled and subgroup analysis for head and neck cancers by years of cigarette smoking. As shown in P = 0.070, P = 0.154, P = 0.960, P = 0.899, P = 0.274, and P = 0.139, respectively.In order to compare the difference and evaluate the sensitivity of the meta-analyses, we repeated the meta- analysis with one study excluded at each time. The results did not have significant change of the pooled OR even if the most influential study was omitted. We also excluded studies by different subgroup , and we did not find any significant change of the pooled OR (data not shown). These indicated the robustness of our findings. Funnel plot and Egger’s test were performed to assess the publication bias. The shape of the funnel plot –S6 Figs The results of the present meta-analysis, based on 13,830 HNC cases, provide quantitative testimony for a positive association between alcohol and tobacco consumption and HNC risks. Compared with non/occasional drinkers, we found that alcohol consumption was associated with a significantly increased risk at moderate and heavy levels for HNC patients, while our results showed much higher ORs of pharynx cancer compared with those presented among patients with oral and larynx cancers. Moreover, our results were rather similar to those shown in previous studies , 47. TheA significantly increased risk at all three different levels was found from the results of the tobacco consumption meta-analysis. It showed a significantly higher increased risk for larynx cancers than oral and pharynx cancer. One of the possible theory was that fibers of tobacco could be released and then penetrated into the laryngeal tissue, thus becoming a further risk factor initiating tumorigenesis . AnotherIn some epidemiologic studies, ethanol had been described as a risk enhancer in smokers, but not as an independent risk factor . HoweverLimitations of our meta-analysis include some possible residual confounding, such as diet, physical activity and human papillomavirus (HPV) infection. But we could not ignore and exclude these residual confounding. Some studies demonstrated that HPV infection is a risk factor for HNC cancer, especially for cancers of the tongue ,55. HoweIn conclusion, tobacco consumption was found to be a stronger risk factor for head and neck cancer than alcohol consumption. The pharynx was the most affected by the harmful effects of alcohol, while larynx was the most affected by the harmful effects of tobacco. Europeans should pay more attention to harmful effects of tobacco. Tobacco consumption increased the risk of head and neck cancer even for smaller quantities whereas alcohol drinking increased this risk significantly at moderate and heavy levels. Precancerous lesions or cancer at an early stage should be detected by regular check-ups. Prevention efforts should be focused on smoking and drinking cessation .S1 Fig(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S3 Fig(TIF)Click here for additional data file.S4 Fig(TIF)Click here for additional data file.S5 Fig(TIF)Click here for additional data file.S6 Fig(TIF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.


Pancreatic cancer ranks as the fourth leading cause of cancer-related mortality in the USA. And gemcitabine has been the standard of care for advanced pancreatic cancer. However, a combined use of gemcitabine plus cisplatin (GemCis) has shown promising efficacies in pancreatic cancer patients. Here, system review and meta-analysis were performed to compare the efficacy and safety of GemCis versus gemcitabine (Gem) alone in the treatment of pancreatic cancer.The databases of MEDLINE (PubMed), EMBASE, and Cochrane Library were systematically searched for retrieving the relevant publications prior to 31 September 2014. The primary end point was overall survival (OS) and secondary end points included 6-month survival, 1 year survival, overall response rate (ORR), clinical benefit rate (CBR), time to progression/progression-free survival (TTP/PFS), and toxicities.P = 0.004), ORR , PFS/TTP (hazard ratio (HR) = 0.87; 95 % CI 0.78–0.93, P = 0.022), and the overall toxicities . However, no significance difference existed in overall survival , 1-year survival rate , and CBR . As for grade III/IV toxicity, seven kinds of toxicities were higher in the GemCis group. However, no significant inter-group statistical differences existed in the incidence of leukopenia, thrombocytopenia, or diarrhea.A total of nine randomized controlled trials involving 1354 patients were included for systematic evaluations. Overall, as compared with Gem alone, GemCis significantly improved the 6-month survival rate (relative risk (RR) = 1.303, 95 % confidence interval (CI) 1.090–1.558, Despite a higher incidence of three-fourths toxicity, GemCis offers better outcomes of ORR, PFS/TTP, and 6-month survival, which indicates GemCis may be a promising therapy for pancreatic cancer. As the fourth leading cause of cancer-related death in the USA , pancreaThe management modalities of pancreatic cancer may be summarized as surgical resection, chemotherapy, radiotherapy, chemoradiation, etc. Despite an availability of different managements, the outcomes of pancreatic cancer patients were similar with a median survival of 20 to 22 months . So imprIn 1997, gemcitabine (Gem) became a de facto standard of care for advanced pancreatic cancer. As compared to 5-fluorouracil, it modestly increased overall survival and offers better clinical benefits . Though As a valid inducer of apoptosis in pancreatic cancer cells, cisplatin is among the most effective and widely used chemotherapeutic agents . A combiA previous meta-analysis presented by Xie et al.  in 2006 From inception to 31 September 2014, comprehensive electronic searches were performed within the database of EMBASE, MEDLINE (PubMed), Cochrane Central Register of Controlled Trials, China National Knowledge Infrastructure (CNKI). The relevant medical subject heading and free words included “gemcitabine,” “gemzar,” “cisplatin,” and “pancreatic cancer/carcinoma/adenocarcinoma.” And the searching languages were limited to English and Chinese.Two authors  independently abstracted the data from each study. Any discrepancies between the two reviewers were resolved by consensus and discussion. The following parameters were extracted: (1) publication and first author’s name; (2) patients, characteristics, number of eligible patients, treatment arm, study design, and follow-ups; (3) treatment outcome, such as OS, 6-month and 1-year survival rates, ORR, CBR, TTP/PFS, and toxicities.The studies included in this meta-analysis should fulfill the following inclusion criteria: (1) cytologically or histologically confirmed advanced stage and/or metastatic pancreatic cancer; (2) baseline Karnofsky performance status score ≥50 % (or ECOG performance status <2) and adequate renal, hematological, hepatic, and cardiac functions; (3) aged over 18 years; and (4) without antitumor therapy within 6 months before study. And the exclusion criteria were studies without a full text and non-published conference abstracts. For duplicated literature reports, the most comprehensive ones were selected. Also, other reports might be supplemented.χ2-based Q-test and I2 statistics were used to assess heterogeneity of studies. If there were statistical differences in terms of heterogeneity , a random effects model was selected; otherwise, a fixed effects model was used. Forest plots for each meta-analysis were used to present the raw data  for each arm of the study potential. Publication biases were ascertained by visually inspecting funnel plots for ORR analysis. The relative symmetry of individual study estimates was assessed around overall estimate, followed by Begg’s and Egger’s tests.All analyses were performed strictly with Stata version 12.0 software . The primary end points included OS. Six-month and 1-year survival rates, ORR, CBR, TTP/PFS, and toxicities were used as second end points. Relative risk (RR) was calculated with a method for dichotomous data and weighted mean difference (WMD) for continuous outcomes and pooled across studies using the DerSimonian and Laird random effects model . The hazA total of 424 potentially relevant publications were identified by initial electronic searches. After duplicating and reviewing the full texts, only nine articles –21 were The basic characteristics of included studies are summarized in Table I2 = 0, P = 0.926) was found; therefore, a fixed model was employed for meta-analysis of HR. The overall meta-analysis revealed that HR was lower for the patients treated with GemCis than with Gem alone. However, no difference was found between the two groups  . And a small significant heterogeneity existed  . However, no significant inter-group heterogeneity existed in the 1-year survival rate   Fig. .Fig. 4OnI2 = 5.2 %, P = 0.390). So, a fixed effects model was employed. Significant differences existed between GemCis and Gem-alone groups. And the meta-analysis revealed that the combination group was linked with higher ORR than the Gem-alone group  . And small heterogeneity was seen between the two groups regarding the outcome of CBR   and Gem alone (n = 330). Overall, there was a significant increase in the TTP/PFS when the GemCis is compared with the Gem group . And the heterogeneity between the two groups regarding the outcome of TTP/PFS was low  .The grade III/IV toxic effects of chemoradiotherapy were summarized in the table. As shown in Table P = 0.71 for ORR, see Fig. P = 0.39 for ORR).Begg’s funnel plot and Egger’s test were performed to assess the publication bias of the selected studies for ORR analysis. The shape of the funnel plots did not show obvious evidence of asymmetry ; the heterogeneity that was not found may result from the same chemotherapy medicine. The TTP/PFS reported by the existing studies varied from 3.6 to 10.4 months in GemCis regimen and from 2.0 to 12.2 months in Gem. The pooled HR for TTP/PFS performed by our analysis was 0.87, indicating a 13 % reduction in the risk of death in patients treated with GemCis regimen. The results were also consistent with those of Heinemann’s study. Comparing GemCis with Gem alone, OS and PFS/TTP of GemCis were both superior to those of Gem alone [In the present meta-analysis, we examined the therapeutic efficacy and safety of Gem-plus-cisplatin-based versus Gem alone for pancreatic cancer. Several clinical studies suggested that a combined use of GemCis was feasible, and it could improve ORR, PFS, and OS , 19–22. em alone . HoweverIn the present study, GemCis significantly increase ORR by 48.2 % as compared with Gem alone. Two recent meta-analyses of Gem-plus-fluorouracil versus Gem alone indicated that the combination group significantly improved the outcome of ORR , 30. AndP = 0.000). This result was in accordance with another meta-analysis comparing Gem-plus-fluorouracil versus Gem alone [In our meta-analysis, we found that GemCis was associated with high incidence of adverse events. Pooled toxicity data significantly increased by 116 % in GemCis versus Gem alone (RR = 2.164, 95 % CI 1.837–2.549, em alone . AlthougSome limitations of the present meta-analysis should be acknowledged. Firstly, this meta-analysis only drew upon published data rather than individual patient profiles. Secondly, the sample size was too small to have a sufficient statistical power for the efficiency and safety of pancreatic cancer between GemCis and Gem. Therefore, more studies with larger sample sizes are needed. Third, different doses of cisplatin yielded divergent outcomes.The present study meta-analysis revealed a significant improvement in the 6-month survival rate, PFS/TTP, and ORR of pancreatic cancer. However, no significant difference existed in OS, 1-year survival, and CBR. The incidence of grade III/IV toxicity was higher for GemCis than for Gem alone. Yet, the incidence of adverse events for GemCis remained generally tolerable. In conclusion, a combined use of Gem and cisplatin is superior to Gem alone as an alternative chemotherapy for pancreatic cancer. However, owing to the above limitations, more convincing studies are warranted.


TCF4 and PTPRG gene variants in Fuchs' corneal dystrophy (FCD).A meta-analysis of To identify novel genetic markers in patients with FCD in different ethnic populations.TCF4 and PTPRG in FCD. Odds ratios (OR) and 95% confidence intervals (CI) of each single-nucleotide polymorphism (SNP) in allelic, dominant and recessive models were estimated using fixed-effect model if 2I<50% in the test for heterogeneity, otherwise the random effects model was used.MEDLINE and EMBASE were searched for eligible genetic studies on TCF4 and two in PTPRG. There were 1610 FCD patients and 1565 controls tested for TCF4 rs613872. This SNP was strongly associated with FCD in Caucasians (P = 5.0×10−106), with the risk allele G conferring an OR of 3.95 (95% CI: 3.49–4.46). A further 4 TCF4 SNPs  were also significantly associated with FCD in Caucasians (P<10−8). However, we found no SNP associated with FCD in Chinese. In addition, there was no significant association between FCD and PTPRG.Thirty-three records were obtained, with 8 being suitable for meta-analysis, which included five SNPs in TCF4 rs613872 is strongly associated with FCD in Caucasians but not in Chinese, which may suggest ethnic diversity in FCD SNP associations. SNPs in PTPRG were not associated with FCD in Caucasians or Chinese populations. Results of this meta-analysis indicate the need for larger-scale and multi-ethnic genetic studies on FCD to further explore the associated gene variants and their roles on the mechanism and genetic basis of FCD. Fuchs' corneal dystrophy (FCD) (MIM 136800) is a bilateral, asymmetric, progressive disorder affecting the corneal endothelium. It is characterized by the formation of guttata which are microscopic outgrowths on Descemet membrane. The resulting thickening of Descemet membrane causes a loss of corneal endothelial cells leading to corneal edema and eventual visual loss FCD1, FCD2, FCD3 and FCD4 on chromosome 13, 18, 5, and 9 respectively ZEB1 located on 10p11.2, SLC4A11 on 20p12 and LOXHD1 on 18q21.1 have been reported as causal genes ZEB1, encoding the Zinc finger E-box-binding homeobox 1 transcription factor, also known as TCF8, harbors the missense mutations p.Q840P, p.N78T, p.P649A, p.Q810P and p.A905T in patients with FCD SLC4A11 that have been identified in both sporadic and familial cases LOXHD1 is located on chromosome 18 (FCD2) and mutations of LOXHD1 result in aggregates in the endothelium and increased thickness with protein abundance of the Descemet membrane which are both pathognomonic of FCD AGBL1 in the 15q locus has been identified in patients with FCD recently and its truncated protein product AGBL1 has altered protein-protein interaction with protein transcription factor 4 (TCF4) COL8A2 on chromosome 1p34.3-p32.3. This type of FCD shows gender differences but still has histopathological characteristics of the endothelial guttata from the classic late-onset form of FCD. It has therefore been regarded as a different form of the same disease The etiology of FCD is not fully understood. Genetic factors have been suggested as a major risk factor TCF4 and rs10490775 in the protein tyrosine phosphatase receptor type G (PTPRG) gene in FCD patients TCF4 gene is located on chromosomal region 18q21. The upstream gene, CCDC68, is over 260 kbps away from TCF4, while the downstream hypothetical gene, FLJ45743, is over 600 kbps away (HapMap Data Release 28). Therefore, rs613872 is less likely to tag another gene in this region. The TCF4 gene encodes the E2-2 protein, which is expressed in the developing corneal epithelium TCF4 can induce epithelial-mesenchymal transition (EMT) in epithelial cells and is vital in corneal damage repair TCF4 rs613872 is within binding site for two transcription factors; Ini1 (SMARCB1) and BRG1 (SMARCA4) which are components of the SWI/SNF chromatin remodeling complex involved in transcriptional regulation TCF4 and PTPRG genes are significantly associated with FCD, these genes also show heterogeneity in the association profiles across populations. Thus the current meta-analysis will aim to give a comprehensive review of all the relevant studies to demonstrate clarity in association of these genes with FCD in different population groups. This is to establish the significance of TCF4 and PTPRG related to FCD, with the highlight of relevant SNPs. Included in this meta-analysis are all the reported associations of SNPs in or near TCF4 and PTPRG with FCD. The association credibility and the effects of the various genes are also evaluated.In 2010 a genome-wide association study (GWAS) identified the single-nucleotide polymorphisms (SNPs) rs613872 in A systematic literature search was conducted in the MEDLINE and EMBASE databases  with the following free words and MeSH terms: “E2-2”, “transcription factor 4”, “TCF-4”, “TCF4”, “protein tyrosine phosphatase, receptor type, G”, “HPTPG”, “PTPG”, “PTPRG” and “Corneal Dystrophies, Hereditary”, and “Fuchs' Endothelial Dystrophy”. We supplemented our search by screening the reference lists of all the relevant studies, including original articles, reviews, and meta-analyses. No language filter was applied.CBM, CNKI, VIP) but did not find any additional relevant studies. For studies published by the same group on the same gene and markers on overlapping sample population only the most recent article or the article with the largest sample size was included for analysis. Data from non-overlapped sample populations from same study are regarded as different sample collections.The following inclusion criteria were applied in the review process: (1) association studies from peer-reviewed journal evaluating the association between variants in selected genes and the disease; (2) genotype or allele counts and/or frequencies in case and control groups were presented, or such data could be calculated from enumerative data provided in the articles; (3) case and control groups were unrelated and drawn from the same temporally and geographically defined population; (4) control subjects are free of any form of corneal dystrophy. All animal studies, case reports, reviews, abstracts, conference proceedings, editorials and reports with incomplete data were excluded. Although non-English articles were not deliberately excluded, the articles in the final analysis were all in English. We also searched some Chinese databases (Two reviewers (LL and LM) separately extracted data from the retrieved records and confirmed the validity of the included articles. Any discrepancy was resolved by other reviewers (SSR and LJC). The following variables were extracted: author, year of publication, ethnicity of study subjects, result of Hardy-Weinberg equilibrium (HWE) test in controls, the numbers of patients and controls, demographic information, and the allele and genotype counts or frequencies of each SNP. The allele or genotype counts were calculated from the frequencies, rounding to the closest integer, in those studies where the counts were not given 2I≤50% or random effects model when 2I>50% Polymorphisms reported in two or more studies were meta-analyzed. Odds ratios (ORs) and 95% confidence interval (CIs) for the tested allele  were calculated with fixed effects model if 2I test. The 2I value was interpreted as of no (0–25%), low (25–50%), moderate (50–75%) and high heterogeneity (75–100%) Z-test. The P values were calculated using the Z scores. An association giving a pooled P value of <0.05 was considered statistically significant. The Funnel plots and Egger's test were used to evaluate potential biases P<0.05, publication bias was assumed to exist.Inter-study heterogeneity was tested with the The workflow and results of the literature review are shown in TCF4 and 2 SNPs (rs7640737 and rs10490775) in PTPRG were tested in at least two sample collections and thus were eligible for meta-analysis.The characteristics of all the included articles are summarized in TCF4 gene, namely rs613872, rs17595731, rs2286812, rs618869 and rs9954153, were meta-analyzed. Notably rs613872 and rs17595731 were non-polymorphic in Chinese Five SNPs in the −106, 2I = 0%; −13, 2I = 0%; 2I = 67%) was detected, which could be explained by the differences in Caucasians and Chinese cohorts. In Chinese, this SNP was not significantly associated with FCD . When the data from the two Chinese cohorts −13, 2I = 0%). SNP rs618869 was tested in two cohorts from the same study by Baratz et al. −14, 2I = 0%; −16, 2I = 0%; TCF4 rs17595731 in the recessive model . Although the associations of FCD with SNPs in the PTPRG gene were indicated, the overall P value did not reach genome-wide significance TCF4 rs613872 with FCD in Caucasians (P = 5×10−106) was obtained and there was no heterogeneity found (2I = 0%). The G allele increases the odds of FCD by nearly 4 folds. In contrast, rs613872 is not associated with FCD in ethnic Chinese TCF4 SNPs  with FCD in Caucasians (P<10−8). Rs2286812 has been studied in two Chinese cohorts but showed no significant association TCF4, SNPs rs10490775 and rs7640737 in PTPRG were not associated with FCD in Caucasians (P>0.1) and has distinct heterogeneity across study populations (2I>80%). Accordingly, TCF4 is the only susceptibility gene that has been confirmed so far for FCD.In a recent genome-wide association study Baratz et al. identified the SNP rs613872 in the TCF4 rs613872 in FCD among the Caucasian populations is confirmed and the G allele is a risk factor. However, TCF4 rs613872 and rs17595731 has been investigated in a total of 91 FCD patients, 42 patients with non-Fuchs' corneal dystrophies, and 612 controls of Chinese origin TCF4 SNP rs17089887, near rs613872, was significantly associated with FCD in Singaporean Chinese, with the allele C conferring a 2.57-fold of increased risk TCF4 gene and the associations were similar TCF4 gene in linkage disequilibrium with both rs613872 and rs17089887 that may be responsible for the association signals detected on these two SNPs.This meta-analysis involves the largest sample size to date with a total of 1610 FCD cases and 1565 controls. The presence of TCF4 rs613872 in Chinese as shown in our meta-analysis. Furthermore, according to the database of the Human Genome Diversity Project TCF4 rs613872 is rare or nonexistent in populations from eastern Asia, Africa, and central and southern America but is more frequent in European, Middle Eastern and southern Asian populations. In order to determine the association of TCF4 rs613872 with FCD, further studies involving larger cohorts are thus needed to correlate the prevalence of FCD and the frequencies of TCF4 polymorphisms in other regions of China and Asia.FCD has reportedly lower prevalence among Chinese than Caucasians, and it accounts for only 3%–4.5% of cases requiring penetrating keratoplasty TCF4 gene is located on 18q21.1, encoding the E2-2 protein, a member of the ubiquitously expressed class I basic helix-loop-helix (bHLH) transcription factors which are necessary in cell growth and differentiation TCF4 rs613872 is within the chromatin immunoprecipitation sequence (ChIP-seq)-purported binding site for two transcription factors: Ini1 (SMARCB1) and BRG1 (SMARCA4)TCF4 through Ini1 (SMARCB1) and BRG1 (SMARCA4) and hence its targets TCF4 rs613872 played a significant role in identifying the potential causative genes and possibly future biomarkers for FCD.The TCF4 and PTPRG in FCD. The study has also revealed the limitations in the current FCD genetic studies. In particular, only a limited number of SNPs were found for this meta-analysis, and the number was even smaller in the Chinese populations making false negative errors likely. The small number of studies with small number of study participants in Chinese and Asian populations and non-polymorphism in the relevant SNPs had limited the generalizability of our results and introduced heterogeneity in the data. Further studies on TCF4 in FCD among different Chinese and Asian populations are needed to confirm the role of this gene. Besides, the homogeneity of the ethnical background in different Caucasian study cohorts may have played a role giving rise to subtle difference in the prevalence and severity of FCD which have not been accounted for in this study. It would be of higher impact and generalizability if more details and various ethnicities in relation to FCD were reported. Also, although a comprehensive search has been employed with different strategies to identify all published studies on TCF4 and PRPRG, relevant studies not meeting the search criteria can be missed. There may be useful results from association studies reported in abstracts, conferences and non-English journals. Thus, there can be selection and publication bias as a result of disproportionate exclusion of negative data, although we detected no significant bias by the Egger's test in the present study. Finally, important covariates like gender, age, severity of FCD, or potential gene-gene and gene-environment interactions could not be included in our analysis because raw allelic and genotypic data were not available. However, alleles are unlikely to display opposite effects based on different covariate. The association between FCD and variants in TCF4 and PTPRG as detected in this study should therefore be genuine.This systematic review and meta-analysis is an overview of the published genetic studies on TCF4, with rs613872 having the strongest effect, to be genetic susceptibility factors for FCD in Caucasians but not Chinese populations. To enhance our understanding of the mechanism of FCD, re-sequencing studies and biological functional studies in multiple ethnic groups are required to ascertain FCD-associated gene variants and to determine their pathophysiological effects.In summary, results of this meta-analysis confirm 5 SNPs in Appendix S1Lists of included studies.(DOC)Click here for additional data file.Appendix S2Lists of excluded studies with reason.(DOC)Click here for additional data file.Checklist S1PRISMA checklist.(DOC)Click here for additional data file.


Previous studies indicated that the single nucleotide polymorphisms (SNPs) in TLR9 gene might be associated with Tuberculosis (TB) risk. However, the results are inconsistent and inconclusive.1745 articles from four databases were involved in our study. A meta-analysis on the associations between the seven polymorphisms and TB risk was carried out by comparison using different genetic models.In this systematic review 8 studies from seven English articles were analyzed. Our results showed that rs352139 is significantly associated with TB risk . In the ethnic subgroup analysis, Indonesians with AA genotype had a decreased susceptibility while Mexicans with GG allele had an increased risk.The meta-analysis indicated that rs352139 polymorphism might be associated with decreased TB risk in Indonesians whereas increased risk in Mexicans. Whether the observed association was due to causal effect needs to be further studied. Mycobacterium tuberculosis (MTB), is one of the most contagious diseases in humans. A recent report (2014) from WHO estimated that one third of the global population has been latently infected without clinical symptoms. In 2012, approximately 8.6 million people worldwide were diagnosed with TB and 1.3 million died from TB [Pulmonary tuberculosis (TB) caused by report 204 from WHreport 204 from WHToll-like receptors (TLRs) are pattern recognizing receptors (PRRs) and play an important role in regulating human’s immune system. The TLRs can recognize pathogen-associated molecular patterns (PAMPs) in both extracellular and intracellular environment. The TLRs have the ability to initiate signaling pathways that are responsible for activating both innate and acquired immune responses as well as the production of inflammatory cytokines. The members of TLRs family have been detected on human plasma membrane TLR1, TLR2, TLR4-6) and in leukocyte endosome TLR3, TLR7-9) [, TLR4-6 , TLR7-9 The gene of toll-like receptor 9 is located on chromosome 3p21.3. The total length of TLR9 gene is approximate 5 kb. Its coding gene has two exons, and the major coding region is in the second exon . PreviouResearches have indicated certain race population with special genotype of TLR9 polymorphism might have higher risk for TB; however the findings are inconsistent and inconclusive. From public health perspective, given the widespread and severe nature of the disease, it is important to understand the association between TLR9 polymorphism and TB risk, so that individuals with higher risk genotype could be identified and receive targeted preventive care. In this study, we conducted a systematic review of current literature on this issue and analysis of the associations between TLR9 polymorphisms and TB risk.We searched the EMBASE, Web of science, PubMed and Chinese National Knowledge Infrastructure (CNKI) databases to identify the publications that reported the association between the TLR9 polymorphism and risk for pulmonary tuberculosis from April, 2004 to July, 2014. The key words used were ‘toll like receptor 9, SNP, lung tuberculosis’, ‘toll like receptor 9, snp, pulmonary tuberculosis’, ‘toll like receptor 9, polymorphism, pulmonary tuberculosis’, or ‘toll like receptor 9, polymorphism, lung tuberculosis’. Only the articles in English or Chinese with an English abstract were selected. After excluding duplicates, titles and abstracts were reviewed. Studies would be selected if they: 1) were case-control studies; 2) reported genotype distribution in both cases and controls. The studies were excluded if they were: 1) review articles; 2) not related to TLR9; 3) not a human study; or 4) not related to the association between tuberculosis and host genetics. The study selection process was summarized in Figure P values for HWE evaluation), source of controls, and genotyping methods .Hardy-Weinberg Equilibrium (HWE) was examined in controls by asymptotic Pearson’s Chi-square test for each polymorphism in each study. The association between polymorphism and TB risk was estimated with odds ratios (OR) and corresponding 95% confidence intervals (CI). Between studies heterogeneity was tested using Q test and Iere used . The pubere used . P < 0.0We used ‘Meta-analysis of Observational Studies in Epidemiology’  as guideA total of 1745 articles were obtained from the literature search. As shown in Figure 2 = 0.0%). The overall OR  using fixed-effect model was 0.93 (95% CI 0.79–1.10), P =0.43. Analyses of other genetic models were also performed, and no association was identified (Table The Human Genome Variation Society (HGVS) name of rs187084 is NM_017442.3:c.-1486 T > C ,20. Thre2 = 0.0%). The overall OR  using fixed-effect model was 1.09 (95% CI 0.90–1.33), P =0.37. Analyses for other genetic models also did not identify significant association . The overall OR  using the random-effect model was 0.98 (95% CI 0.77–1.25), P =0.87. Analyses for other genetic models did not identify significant association  and TB risk. The analysis revealed an association between certain TLR9 polymorphism and TB risk. In addition, 5 different genetic models  were analyzed in all polymorphisms. A subgroup analysis by race was also performed for rs187084, rs352139 and rs5743836 polymorphisms, the studies of which included Indians, Iranian and West African, Indonesians, Vietnamese, Chinese and Mexicans. The results showed that rs187084 and rs5743836 polymorphisms were not associated with TB risk, while the association between rs352139 polymorphism and TB risk may vary by race.Mycobacterium Tuberculosis required TLR9 in animal model [We chose TLR9 gene in our analysis because it’s critical role in the incidence of TB from previous studies. Chen et al. demonstrated that cell apoptosis induced by al model . Anotheral model . All theTLR9 is a recognition receptor in leukocytes endosome, especially in B cells, plasmacytoid dendritic cells (pDCs) and macrophages. It is localized in the endosomal and lysosomal compartments. It is one of the most important receptors for the initiation of innate immune response against intracellular pathogens. The function of TLR9 is to recognize the unmethylated Cytosine-phosphate-Guanine (CpG) sequence which usually appears in bacterial DNA . In receSNPs rs352165 and rs352167 were located in intron 5 of aminolevulinate delta-, synthase 1 (ALAS1) gene in the downstream of TLR9. Sanchez D et al. have reported that these two SNPs were in linkage disequilibrium (LD) with rs352139 and rs352140 SNPs of TLR9 . Thus thIn our study, the association of rs352139 with TB risk varied by race. In Indonesians AA genotype is associated with lower risk of TB, while in Mexicans it is associated with higher risk of TB. The Mexicans with AA allele had 2.73 times higher risk of TB than Indonesians with the same genotype. There is no known explanation for the racial difference. Also, we observed a significant association between rs352139 polymorphism and TB risk in Indonesians for Allele, Heterozygote and Homozygote models. There was also a significant association between rs352139 polymorphism and TB risk in Mexicans for Allele, Dominant and Recessive models. Although 4 case-control studies of rs352139 were included in this analysis, there was only one study for each racial group. Therefore, our results should be interpreted with caution, and more studies with larger sample size should be considered to confirm this differential risk by race.The SNP rs352139 is located in the intron of TLR9 gene (Figure The SNP rs187084 is located in the promoter of TLR9 gene, which could create a Sp1 binding site . VariantThe SNP rs5743836 is located in the promoter region of TLR9. This SNP was thought to be associated with increased transcriptional activity and function of TLR9 . In our This systematic review summarized the current literature on the association between TLR9 polymorphisms and TB risk. Our results indicated that rs187084, rs352165, rs5743836, rs5743842, rs352140 and rs352167 polymorphisms did not show significant association with TB risk. The rs352139 polymorphism might be associated with decreased TB risk in Indonesians whereas increased risk in Mexicans. However, the data were based on single study for each race. Whether the observed association was due to causal effect needs to be further studied.