ESR1) play critical roles in breast cancer (BC) susceptibility. Genome-wide association studies have reported that SNPs in ESR1 are associated with BC susceptibility; however, the results of recent studies have been inconsistent. Therefore, we performed this meta-analysis to obtain more accurate and credible results.Single nucleotide polymorphisms (SNPs) in the estrogen receptor gene  with 95% confidence intervals (CIs) to assess the strength of associations using fixed effects models and random effects models. Twenty relevant case-control and cohort studies of the 3 related SNPs were identified.ESR1 gene, rs2077647:T>C, rs2228480:G>A and rs3798577:T>C, were not associated with increased BC risk in our overall meta-analysis. Stratified analysis by ethnicity showed that in Caucasians, the rs2228480 AA genotype was associated with a 26% decreased risk of BC compared with the GG genotype . The C allele of the rs3798577:T>C variant was associated with decreased BC risk in Asians , while Caucasians with this allele were found to experience significantly increased BC risk . A non-significant association between rs2077647 and BC risk was identified in all of the evaluated ethnic populations.Three SNPs of the Rs3798577 was associated with an increased risk of BC in Caucasian populations but a decreased risk in Asians. Rs2228480 had a large protective effect in Caucasians, while rs2077647 was not associated with BC risk. PreviouESR1 gene is a steroid hormone receptor gene located on chromosome 6 at 6q25.1. It contains eight exons spanning ~295 kb [ESR1 gene were shown to be associated with BC risk, including rs2234693, rs1801132, rs9340799, rs2077647, rs2228480 and rs3798577, and also, studies have showed that the genetic variants played important roles in the transcription and protein expression[ESR1 genetic variant and BC risk. One study showed a protective effect of rs2077647 on BC risk, another study reported that ESR1 rs2077647 increased BC risk, and the remaining studies failed to replicate these associations. Three studies showed that the rs3978577 SNP, which is located in the 3’ UTR of ER-α, increased the overall risk of BC, one study provided evidence that it decreased BC risk, and the others also failed to replicate these associations.The human  ~295 kb . Many SNxpression. Recentlxpression, while txpression–34. ThisESR1 polymorphisms and the risk of BC, a new meta-analysis that integrated more recent studies with earlier publications was conducted.Although rs3798577 and rs2228480 were discussed in a meta-analysis in 2010, the analysis included only 4 studies for each SNP . HoweverESR1” or “Estrogen receptor α” or “ER alpha” or “Estrogen receptor alpha” or “ERα”). Google Scholar was also used to search for relevant studies. Chinese papers were selected by searching the WanFang Data, Chongqing VIP (CQVIP), and China National Knowledge Infrastructure (CNKI) databases using the same search terms. The references of eligible articles were also inspected to find other potential studies. Only studies published in English or Chinese were included in this meta-analysis; any disagreement was resolved via discussion between two of the authors (H.H. and J.Z.). E-mail was used to contact study authors to obtain full text articles or missing data. This study was performed in accordance with the PRISMA statement checklist  and (“breast cancer” or “breast neoplasm” or “carcinogenesis” or “breast carcinoma” or “breast tumor” or “BC” or “mammary cancer”) and  availability of odds ratios (ORs) with 95% confidence intervals for polymorphisms and haplotypes or sufficient genotyping data to estimate these parameters; and (3) all diagnoses of BC confirmed by pathological or histological examination. Reviews, simple commentaries, case reports and meta-analyses were excluded. For overlapping studies, only the study with the largest sample was included.The inclusion criteria were the following: (1) case-control or cohort study focused on associations between The data from the published studies were extracted independently by two of the authors, and consensus was reached on all of the items. For each study, the following variables were collected: first author’s name or study organization name, year of publication, area, language, ethnicity, study methods, number of cases and controls, sources of cases and controls, allele and genotype frequencies, Hardy-Weinberg equilibrium (HWE), OR value, statistical power and minor allele frequency (MAF) in the controls. OR adjustment factors are not listed in our tables because every study used different factors for OR adjustment; therefore, it was difficult to find common factors for our meta-analysis.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). The NOS contains two different quality assessment scales for case-control studies and cohort studies. The two different forms each consist of three groupings, but the grouping items differ. The NOS identifies “high”-quality choices with a “star”, with a maximum of one “star” for each item within the “Selection” and “Exposure/Outcome” categories, and a maximum of two “stars” for “Comparability”. To obtain objective outcomes, any disagreement was discussed, and another author was consulted.The Newcastle-Ottawa Quality Assessment Scale (NOS)  was usedESR1 polymorphisms with BC susceptibility was measured by ORs with 95% CIs in four genetic models, including a variant heterozygote versus wild-type homozygote model, a variant homozygote versus wild-type homozygote model, a dominant model, and a recessive model. Between-study heterogeneities were estimated using the χ2-based Q test [P<0.05. The I2 statistic was then used to quantitatively evaluate heterogeneity  [P<0.05) or I2>50% indicated heterogeneity among the studies, a random effects model (DerSimonian Laird method) was used to conduct the meta-analysis; otherwise, a fixed effects model (Mantel-Haenszel method) was used. To explore the sources of cross-study heterogeneity, subgroup analysis by ethnicity was performed. HWE of the genotype frequencies in the control group was assessed by the goodness-of-fit χ2 test. Sensitivity was evaluated by omitting each study one at a time to assess the influence of each study on the overall estimate [fs) was also used to assess the stability of the results through comparison with the number of relevant included studies. All of the P values were two sided, with significance defined at 0.05. All analyses were performed using Review Manager software . The gene data for the heterogeneity analysis were download from the International HapMap Project (http://hapmap.ncbi.nlm.nih.gov/). Allele frequencies for the three polymorphisms in different populations were assessed by the goodness-of-fit χ2 test, and the linkage disequilibrium (LD) analysis was performed using Haploview software (version 4.0).The association of the d Q test , and thegeneity) . When a estimate . Publicaestimate , 39. TheThe initial search of EMBASE, PubMed, and Web of Science yielded 1184 relevant articles, and an additional 24 records were identified through other sources. Following the deletion of duplicate results obtained from multiple databases, 368 records remained. After the titles and abstracts of the 368 articles were reviewed, 47 full-text articles were finally considered eligible. Ultimately, 20 eligible studies –34 were The characteristics of the 20 eligible studies are presented in The evaluation of the associations of these three polymorphisms with BC risk and the stratified analyses by ethnicity are presented in P value for heterogeneity was less than 0.05 in the dominant model and variant heterozygote versus wild-type homozygote model; therefore, the ORs were pooled in a random effects model. No significant association was found between the rs2228480 genetic variant and BC in any of the four models, and no significant effect was found in Asians. However, Caucasians carrying the rs2228480 TT genotype had a 26% decreased risk of BC compared with those with the CC genotype    Fig 2)..P value P values less than <0.05, which were considered to be statistically significant. For rs2077647, the eligible studies included 6037 BC patients and 7385 control subjects. In the overall population, the Q test of heterogeneity was significant in the variant homozygote versus wild-type homozygote model, and the analysis was conducted using random effect models. There was no obvious association between the SNP and BC risk in any of the genetic models. The subgroup analysis revealed similar results in the Asian, Caucasian and mixed ethnic groups ..P valuesP = 0.019, Nfs = 11; CC vs. TT: OR = 0.72, 95% CI: 0.61–0.85, P = 0.000, Nfs = 23; (CT+CC) vs. TT: OR = 0.78, 95% CI: 0.69–0.88, P = 0.000, Nfs = 29; CC vs. (TT+CT): OR = 0.80, 95% CI: 0.70–0.93, P = 0.003, Nfs = 11). In the dominant, recessive and variant homozygote versus wild-type homozygote models, Caucasians carrying the variant C allele were found to experience significantly increased BC risk  vs. TT: OR = 1.39, 95% CI: 1.01–1.91, P = 0.041, Nfs = 23; CC vs. (TT+CT): OR = 1.21, 95% CI: 1.00–1.47, P = 0.050, Nfs = 8). However, no significant associations were found in the mixed population. The data are presented in detail in For rs3798577, the eligible studies included 8140 BC patients and 10386 control subjects. In the overall population, there was significant heterogeneity in all of the genetic models, so the analysis was conducted using random effect models. We failed to find a significant main effect on BC risk in any of the test models. In the ethnicity subgroup analysis, we found that among Asians, the variant C allele was associated with a decreased BC risk in all of the genetic models  also showed that there was no evidence of publication bias for any of the three polymorphisms .Funnel plots and Egger’s tests were used to assess the publication bias of the included studies. The funnel plots did not reveal any evidence of obvious asymmetry in the three SNPs in the variant homozygote versus wild-type homozygote model . Egger’sfs of the positive result indicated that the results in this meta-analysis were also relatively stable and credible.Sensitivity analyses were performed to evaluate the effect of each study on the pooled ORs through sequential removal of individual studies . No indiHeterogeneity analyses were performed to explore the reason for the heterogeneity in the associations found in the Caucasian and Asian populations. Measures of LD and allele frequencies for the three polymorphisms in the different populations comprised the two parts of this analysis.P values for χ2 test >0.05) showed that there was no heterogeneity in the allele frequencies for the three polymorphisms in the different populations .Allele frequencies for the three polymorphisms in the different populations are listed in The LD plots of all SNPs that were previously found to be associated with BC in different populations are presented in ESR1 gene have been shown to alter ER-α expression and to therefore modulate downstream signaling and BC susceptibility [ESR1 gene plays an important role in the progression of breast carcinogenesis by inducing cell proliferation, programming cell death and accumulating genetic mutations [ESR1 gene that are correlated with susceptibility have been identified.Genetic variants in the tibility . The ESRutations . Many geOur findings showed that the SNPs rs2077647, rs2228480 and rs3798577 were not associated with BC risk in the four test models included in our overall meta-analysis. After the data were stratified by ethnicity, the analysis demonstrated that rs3798577 was associated with an increased risk of BC in Caucasians but had a protective effect in Asians. SNP rs2228480 also had a significant association with BC risk in Caucasians. The strength of the association of rs2228480 and rs3798577 with BC risk varied greatly across ethnic groups. An earlier study  indicateComparison of allele frequencies showed that there were not heterogeneous among the different populations, but the LD plots for the rs3798577 in the different populations showed an opposite result. Hence, two potential reasons for the reversed interaction in rs3798577 between the different ethnic populations can be proposed. First, it may be caused by the differences in the function of genetic variants among different ethnic populations. Second, heterogeneity in LD for the rs3798577 in the different populations is also the possible reason.ESR1 region associated with BC have been reported [GWAS have provided a powerful approach for identifying common disease alleles. Recent GWAS have identified several genetic susceptibility loci for BC, and low-penetrance variants in the reported –46. For reported –49. Our ESR1 mRNA or mRNA expression were detected. Furthermore, another study [ESR1 rs2077647:T>C is a silent coding polymorphism located in exon 1, it is unlikely to alter the protein encoded by ESR1. One research [For rs2077647:T>C, on the one hand, some studies , 23, 40 research  indicateESR1. Although the underlying biological mechanism and its functionality are not yet known, one plausible hypothesis is that rs3798577 polymorphisms might be major regulators of ER-α expression and might modify mRNA stability and ESR1 gene expression.The rs3798577:T>C polymorphism is located in the 3’ UTR of ESR1 and a synonymous variant. The functionality of this SNP is not yet known, but it seems to act as a regulator. Exon 8 is involved in the assembly of the C-terminal region of ER-α, which contributes to the regulation of reciprocal action between ER-α and other transcription factors [The rs2228480:G>A polymorphism is a silent polymorphism located in exon 8 of  factors . Althoug factors , rs22284ESR1 rs3798577:T>C polymorphism was associated with BC risk in a manner that depended on patient ethnicity.The present study had several strengths. Most importantly, it was the first meta-analysis conducted to evaluate the association between rs2077647 and BC risk. It was also the biggest and most recent meta-analysis of the association of rs2228480 and rs3798577 with BC risk, and it was more powerful than previous cohort and case-control studies. In addition, a subgroup analysis was conducted and demonstrated that the However, some limitations of this meta-analysis must be addressed. First, the sample size was relatively small for stratified analyses and might not have provided sufficient power to estimate the associations. Second, the overall OR was based on individual unadjusted ORs, and some important confounding factors, such as age, sex, menopausal status, and BMI, must be adjusted for. Finally, although the funnel plots and Egger’s tests showed that publication bias did not affect our results, only studies published in English or Chinese were included, which produced selection bias at the start of our study.ESR1 rs3798577:T>C polymorphism might be a risk factor for BC in Asians and that the ESR1 rs3798577:T>C polymorphism and ESR1 rs2228480:A>G polymorphism had a large protective effect in Caucasians, while the ESR1 rs2077647:T>C polymorphism was not associated with BC risk. However, the functions of these SNP gene variants in the development of BC and the full mechanisms underlying their effects are still unclear. In the future, more comprehensive and well-designed studies should be conducted to re-evaluate the associations of these three SNPs and other ESR1 gene polymorphisms with BC risk.In conclusion, our meta-analysis indicated that the S1 PRISMA Checklist(DOCX)Click here for additional data file.S2 Checklist(DOCX)Click here for additional data file.S1 Fig(PDF)Click here for additional data file.S1 Table(XLSX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(XLSX)Click here for additional data file.S4 Table(XLSX)Click here for additional data file.S5 Table(XLSX)Click here for additional data file.S6 Table(XLSX)Click here for additional data file.S7 Table(DOCX)Click here for additional data file.S1 Text(DOCX)Click here for additional data file.S2 Text(DOCX)Click here for additional data file.


Epidemiological studies have shown that E-selectin gene polymorphisms (A561C and C1839T) may be associated with essential hypertension (EH), but the results are conflicting in different ethnic populations. Thus, we performed this meta-analysis to investigate a more authentic association between E-selectin gene polymorphisms and the risk of EH.We searched the relevant studies for the present meta-analysis from the following electronic databases: PubMed, Embase, Cochrane Library, Google Scholar, Web of Science, Wanfang Data, and China National Knowledge Infrastructure (CNKI). Odds ratios (OR) with 95% confidence interval (CI) were used to evaluate the strength of the association between E-selectin gene polymorphisms and EH susceptibility. The pooled ORs were performed for dominant model, allelic model and recessive model. The publication bias was examined by Begg’s funnel plots and Egger’s test.P<0.001) in dominant model, 5.284  in recessive model and 2.359  in allelic model. Four studies (six cohorts) evaluated C1839T polymorphism and EH risk, including 1,700 cases and 1,681 controls. The pooled OR was 0.785  in dominant model, 1.250  in recessive model and 0.805  in allelic model.A total of eleven studies met the inclusion criteria. All studies came from Asians. Ten studies (12 cohorts) evaluated the A561C polymorphism and EH risk, including 2,813 cases and 2,817 controls. The pooled OR was 2.280 (95%CI: 1.893–2.748, The current meta-analysis concludes that the C allele of E-selectin A561C gene polymorphism might increase the EH risk in Asian population, whereas the T allele of E-selectin C1839T gene polymorphism might decrease the EH risk. Essential hypertension (EH) is a complex multifactorial disease caused by genetic and environmental factors E-selectin, a cell-surface single chain glycoprotein, was firstly identified in 1985. It is a member of adhesion molecule, also known as endothelial leukocyte adhesion molecule-1(ELAM-1), CD62 antigen-like family member E (CD62E) and SELE. E-selectin is expressed only by activated endothelial cells, and is different from other selectins. The concentration of E-selectin is very low in resting endothelial cells. When vascular endothelial cells are stimulated by inflammatory factors, the expression of E-selectin is greatly increased. Previous researches have revealed that the level of E-selectin is significantly higher in EH group than in controls et al.The gene encoding E-selectin is located in human chromosome 1q22-q25 and contains 14 exons spanning about 13 kilobases of the human genome. Several E-selectin gene polymorphisms have been identified recently. In 1994, Wenzel http://www.wanfangdata.com.cn), and China National Knowledge Infrastructure (CNKI). The following search terms were used in the electronic searches: “E-selectin or endothelial leukocyte adhesion molecule-1 or ELAM-1, or CD62 antigen-like family member E or CD62E” and “variant or mutation or gene or polymorphism” and “essential hypertension”. We also checked the references of relevant studies to minimize the omissions.We searched the relevant studies for the present meta-analysis  from the following electronic databases: PubMed, Embase, Cochrane Library, Google Scholar, Web of Science, Wanfang Data  case-control study evaluating the relationship between E-selectin polymorphism and EH; (b) having the clear original data of genotypic and/or allelic frequencies; (c) papers must be written in either Chinese or English; (d) no restriction on the sample size; (e) hypertension was defined as systolic blood pressure ≥140 mmHg and/or diastolic blood pressure ≥90 mmHg or treatment with antihypertensive medication, and only essential hypertension was included (secondary hypertension excluded). If we were able to obtain several similar data from one research center, we retained the most comprehensive study.A standard protocol was used to record the original data. All relevant studies were read carefully by two investigators (Cai and Zhang) and the original data were extracted independently. If there was any doubtful point, the disagreement was resolved by discussion between the two authors. The characteristics of each study were extracted, including the name of the first author, year of publication, average age, gender, region, ethnicity, numbers of cases and controls, numbers of genotypes in cases and controls, diagnostic criteria of EH and the genotyping methods. If the study did not offer the original data which we need, we tried our best to contact the correspondent author by telephone or E-mail.P = 0.035) in dominant model and 0.805  in allelic model (P = 0.027) in dominant model and 0.769  in allelic model.For C1839T polymorphism, the T allele was found to have a significantly protective effect against EH risk. The pooled OR was 0.785  was adopted to calculate the pooled results if the heterogeneity was not significant . Otherwise, the random-effect model (the Dersimonian-Laird method) was applied Hardy-Weinberg equilibrium (HWE) for the E-selectin genotype distributions of control groups was determined by the Fisher’s exact test. Pooled odds ratios (OR) with 95% confidence interval (CI) were used to evaluate the strength of association between the E-selectin gene polymorphisms and the EH susceptibility. We calculated the pooled ORs by dominant model , allelic model  and recessive model . The subgroup analyses were carried out by ethnicity (Han or others), total sample size (less than 600 or more than 600) and genotyping methods (PCR-RFLP or Taqman PCR). The Q-test and P value <0.05 (two-sided) was considered statistically significant.All statistical analyses were performed by using STATA software version 12.0 for Windows . A Through the literature search, 72 eligible papers were identified, of which 61 papers were excluded. Of the 61 excluded studies, three papers were reviews, two papers were meta-analysis, four studies repeated prior researches, three studies lacked of reported data and failed to obtain relevant information, 49 studies were unrelated to the E-selectin A561C or C1839T gene polymorphisms and EH. The flow diagram of articles selection process is listed in P<0.001) in dominant model, 5.284  in recessive model and 2.359  in allelic model. The E-selectin A561C polymorphism significantly increased the EH susceptibility . Whereas, no significant publication bias was found in the dominant model (P = 0.054) and in recessive model (P = 0.974). For C1839T, there was no significant publication bias according to the visual assessment of funnel plot and Egger’s test .Both Begg’s funnel plot and Egger’s test were performed to evaluate the publication bias of the literatures. To our knowledge, the current meta-analysis was the first study to investigate the relationship between E-selectin gene polymorphisms and EH risk. Our study suggests that E-selectin gene A561C and C1839T polymorphisms might be associated with EH risk.CHD and some type of IS are both characterized by chronic inflammation. EH also has a close association with inflammation. Studies show that several genetic polymorphisms are related to EH and genetic factors maybe account for 30% to 50% of the causes of the variation in blood pressure et al.et al.In 2011, Sun et al. et al. et al. In 2012, Srivastava et al. 2.Compared with A561C polymorphism, the number of study about the connection between E-selectin C1839T polymorphism and EH was relatively small. In 2004, Marteaua et al. et al. By contrast, Li P<0.001, in dominant model). For C1839T polymorphism, the present meta-analysis suggested that the T allele of E-selectin C1839T gene polymorphism might decrease the EH risk .A total of 11 studies included in our meta-analysis. Ten studies (twelve cohorts) evaluated A561C polymorphism polymorphism and EH risk. The C allele of E-selectin A561C gene might increase the EH risk in Asian population . The total sample size of each study was less 1000. Especially, the number of studies on C1839T polymorphism and EH was only 4 (6 cohorts). Secondly, E-selectin was affected not only by the E-selectin gene polymorphisms but also by environmental factors, for example inflammation. Thirdly, all the studies were conducted in Asian populations (most of them were Chinese population). Studies from other continents should be performed in the future. Most importantly, publication bias existed in the current meta-analysis (for EL A561C gene polymorphism). For example, studies may not have been published, if they failed to demonstrate an association between E-selectin A561C gene polymorphism and EH risk. The publication bias could cause the false positive results. In addition, our study only focused on papers published in the English or Chinese, and therefore some eligible studies written in other languages were not included.Despite the limitations, the current meta-analysis concluded that the C allele of E-selectin A561C gene polymorphism might increase the EH risk in Asian populations, whereas T allele of E-selectin C1839T gene polymorphism might decrease the EH risk. Given the above limitations, further studies should be performed to clarify the association between the E-selectin gene polymorphisms and EH through multicenter, large-scale, and multi-ethnic studies in the future.Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.


VEGF rs2146323 polymorphism with the risk of DR. However, the results of these studies are inconsistent. A meta-analysis was performed to evaluate the association between the VEGF rs2146323 polymorphism and the risk of DR.Vascular endothelial growth factor (VEGF) is thought to play an important role in the pathogenesis of diabetic retinopathy (DR). Previous studies have associated the 2 statistic based on the Q test.The PubMed, EMBASE, Web of Science and Google Scholar literature databases until March 2015 were searched. The differences in the studies were expressed in the form of an odds ratio (OR) and the corresponding 95 % confidence interval (CI). Heterogeneity among the studies was tested using the IP = 0.005) and the co-dominant model (CA VS. CC) .A total of four studies (598 cases and 709 controls) were included in the meta-analysis. A significant association was found involving the rs2146323 polymorphism in the dominant model (CA + AA VS. CC) (OR = 1.38, CI = 1.10–1.72, VEGF rs2146323 polymorphism and the risk of DR.Our meta-analysis confirmed the association between the  Diabetic retinopathy (DR) is considered one of the most common microvascular complications in diabetes. It leads to blindness among adults aged 20–74 years and is regarded as the main cause of blindness in people with diabetes . AlthougVEGF is expressed and secreted from vascular endothelial cells , Müller VEGF gene is located on chromosome 6p21.3 with 7 introns and 8 exons and shows polymorphism  or “macular edema/genetics”[mesh]). We chose the literature that involved the polymorphism based on the title and abstract of the articles and even the full text when necessary. The meta-analysis was performed for cases with any form of DR compared with diabetic without retinopathy (DWR). After removal of the duplicates, we used the following inclusion criteria to filter the articles: (1) the study was a case–control study; (2) definitive methods were applied to the diagnosis of diabetic retinopathy, including the Early Treatment of Diabetic Retinopathy Study (ETDRS) and the guidelines of the Expert Committee Report of the American Diabetes Association (DR includes proliferative diabetic retinopathy (PDR) and non-proliferative diabetic retinopathy (NPDR)); (3) the VEGF rs2146323 polymorphism was evaluated; (4) the sample size was sufficient for the calculation of the odds ratio (OR) with 95 % confidence interval (CI); and (5) the control group followed Hardy-Weinberg equilibrium  the name of the first author; (2) the year of publication; (3) the origin country; (4) the racial descent of the study population; (5) the sample size of the cases and controls; (6) duration of diabetes; (7) gender ration ; (8) HbA1c level; (9) the genotype frequency; (10) the genotyping methods; and (11) the HWE  . The New2 statistic based on Q test. Studies with I2 < 50 % or P > 0.10 were considered to be of low heterogeneity, and the fixed effects model was used to calculated the pooled OR, otherwise, random effects model was used to pool OR. STATA software  was used for the analyses. Two-sided P < 0.05 was considered statistically significant.The six genetics models were used for analyses: dominant model: CA + AA vs. CC; allele model: A vs. C; recessive model: AA vs. CA + CC; co-dominant model: CA vs. CC; co-dominant model means AA vs. CC; over-dominant model means CA vs. CC + AA; additive model: CC vs. AA . The difVEGF polymorphism in DR, and 32 articles were irrelevant regarding the VEGF rs2146323 polymorphism were excluded based on reading the titles or abstracts. In addition, 2 articles were excluded because one was not a case–control study [A total of 2724 articles were retrieved in the initial search. Of these, 207 articles were written in a non-English language, 2479 articles were irrelevant regarding the A total of 598 diabetes with retinopathy (DR) cases and 709 diabetes without retinopathy (DWR) controls were identified Table , and 6 gP = 0.308, Egger’s test P = 0.083; allele model: Begg’ test P = 0.308, Egger’s test P = 0.145; recessive model: Begg’ test P = 0.734, Egger’s test P = 0.387; co-dominant model CA VS CC: Begg’ test P = 1.000, Egger’s test P = 0.968; co-dominant model AA VS CC: Begg’ test P = 0.734, Egger’s test P = 0.272; over-dominant model: Begg’ test P = 0.734, Egger’s test P = 0.595; additive model: Begg’ test P = 0.734, Egger’s test P = 0.272). We conducted the sensitivity analysis, and results show that the pooled OR was close to the total pooled OR and was within the total 95 % CI, which indicated that the results of sensitive analyses for these studies were reliable . Fourth, the present studies did not provide the physical and biochemical examination results, which are useful for interpreting the mechanism of the polymorphism in DR. Finally, although DR cases were classified by fundus photographs or OCT, level of DR has been not presented, which may be the source of heterogeneity.In conclusion, our meta-analysis results indicated that there was a significant association between the VEGF rs2146323 polymorphism and the risk of DR. Since most of these polymorphisms are in strong linkage disequilibrium, genome-wide association study (GWAS) can provide more genome-wide genetic association with DR.


There is ample evidence on the effects of prices and taxes on heavy drinking, including binge drinking . ExperimNevertheless, in an issue of Health Economics Review, Dr. Jon Nelson published a “systematic” review article with the following conclusion: “Increased alcohol taxes or prices are unlikely to be effective as a means to reduce binge drinking, regardless of gender or age group.” . We haveThere is a general consensus that the scientifically recommended approach to estimate an overall effect across multiple studies is meta-analysis. Meta-analysis is the final step in the standardized protocol detailed in Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA), which was cited in the Nelson paper. However, instead of conducting a rigorous meta-analysis among the 56 studies that he manually selected and screened, Dr. Nelson deviated from the PRISMA approach by simply counting the number of studies with p ≤ 0.05. Dr. Nelson’s approach failed to account for the heterogeneity of the effect sizes and sample variation due to different studies, and potential publication bias. Dr. Nelson stated that due to “diversity of models and results,” quantitative coefficient estimates for a meta-analysis were not collected. Dr. Nelson’s reference to “diversity of models and results” is the exact reason why quantitative meta-analysis is necessary to obtain a common effect estimate in the first place.Furthermore, among the exclusion criteria, Dr. Nelson stated that studies were excluded if they are “based on a laboratory experiment” or use an interrupted time-series analysis. He did not explain why these two research designs were excluded. Not only do they have better internal validity than other designs, interrupted time series analysis is an ideal method for evaluating natural experiments, a type of study included in Dr. Nelson’s paper and stated in the article title.Based on his characterization of our own work, there is considerable evidence of misconstrued presentation of the literature in Dr. Nelson’s article. For example, Dr. Nelson cited a study by Xuan et al.  and statDr. Nelson further asserted that evidence in several widely-cited studies is drawn from aggregate econometric studies, and here he cited another one of our studies . His asDr. Nelson subsequently stated that “price and tax elasticity estimates for general populations may not apply equally to binge drinkers and other excessive drinkers.” To support this statement, Dr. Nelson cited Ayyagari and colleagues , which wDr. Nelson disclosed that “Research leading to this paper was supported in part by the International Center for Alcohol Policies, Washington, DC.” However, the author failed to disclose his role as a consultant (2012–2015) and member of the Research Advisory Board of ICAP, an alcohol industry-funded “social aspect organization”. ICAP-sponsored research has been criticized for faulty methodology and erroneous interpretations . Nor doeIn summary, Dr. Nelson’s conclusion that taxes were unlikely to be effective at reducing binge drinking is based on faulty methodology and an incorrect interpretation of the findings from studies, and his conclusion contradicts the extensive scientific literature. Binge drinking causes more than half of nearly 80,000 alcohol-attributable deaths in the U.S. each year and accounts for three-quarters of the 224 billion dollars in annual economic costs . This is


P value and associated 95% confidence intervals for the highest versus lowest dietary intake of citrus fruit level. The association was quantified using meta-analysis of standard errors with a random-effects model. Thirteen case–control studies and 6 cohort studies were eligible for inclusion. Citrus intake may significantly reduce risk of esophageal cancer , without notable publication bias  and with significant heterogeneity across studies (I2 = 52%). The results from epidemiologic studies suggest an inverse association between citrus fruit intake and esophageal cancer risk. The significant effect is consistent between case–control and cohort studies. Larger prospective studies with rigorous methodology should be considered to validate the association between citrus fruits and esophageal cancer.Many epidemiologic studies indicate a potential association between fruit and vegetable intake and various cancers. The purpose of this meta-analysis is to investigate the association between citrus fruit intake and esophageal cancer risk. The authors conducted a comprehensive search on PubMed, EMBASE, and the Cochrane Library from inception until July 2014. Studies presenting information about citrus intake and esophageal cancer were analyzed. The authors extracted the categories of citrus intake, study-specific odds ratio or relative risk, and the  The study by Coleman et al19 suggested that dietary fiber may protect against esophageal carcinogenesis, especially esophageal adenocarcinoma. Zhu et al20 found that meat consumption is associated with the risk of esophageal cancer. The intake of red meat is likely to increase the esophageal SCC risk and the processed meat may increase esophageal adenocarcinoma risk; however, the consumption of fish may not be associated with esophageal cancer incidence. This phenomenon may be explained by the effects of various micronutrients such as folate, B vitamins, antioxidants, lutein, and carotenoids.24Cigarettes, red meat, alcohol, hot tea, pickled vegetables, low intake of fresh fruits and vegetables, and low socioeconomic status are associated with a higher risk of SCC.26 which have protective effects against cancer. Previous studies have suggested that citrus intake may improve the incidence of various cancers including pancreatic, breast, and prostate cancers.29 Consequently, we hypothesize that citrus intake is associated with a reduced risk of esophageal cancer. Epidemiologic evidence from cohort and case–control studies on this association has not yet been summarized. Therefore, we conducted a meta-analysis to explore this hypothesis.Citrus fruits include oranges, tangerines, grapefruits, lemons, and limes. They include several components, including flavonoids, folate, carotenoids, and vitamin C,A computerized search of the English language literature on citrus fruits and esophageal cancer yielded no relevant publications from inception to July 2014. We, therefore, decided to use the key words “fruit” and “citrus.” The search terms were  AND ([cancer] OR [tumor] OR [carcinoma]) AND (‘citrus’itrus OR ‘fruit’ruits). We limited the search to human adults without language restrictions. We searched the 3 major electronic databases: PubMed, EMBASE, and The Cochrane Library. Additionally, we reviewed the references from retrieved articles for additional studies. Furthermore, ethical approval was not necessary because our article is a review.29 had to be epidemiologic studies such as case–control and cohort studies. The studies concerning human that addressed the association between citrus intake and incidence of esophageal cancer were collected; however, if the study provides no original data or insufficient information on the odds ratio (OR) or relative risk (RR), and their corresponding 95% confidence intervals (CIs), we excluded it. The studies not measuring the intake of citrus fruits or citrus juice at the individual level are not eligible. The instrument of assessment of citrus intake is questionnaire. Two independent reviewers read the abstracts or full-text articles to assess the eligibility of studies in a standardized manner. We resolved the disagreement by consensus.The included studiesP values were also extracted by us. If separate researches based on the same population were published, we selected the article containing more complete information for inclusion.We extracted important information from all eligible studies. They included study design, country of origin, years of publication, origin of control, number of cases and control, sex distribution, types of citrus fruits, types of cancer, comparison of exposure level, and potential confounding variables adjusted. The estimates of OR/RR, their associated 95% CIs, and 30 of RevMan 5.0. For studies that provided OR/RR by cancer subtypes,31 we used a random-effects model to obtain a pooled estimate from the individual study  of the log OR/RR by using the following equation: SE = (ln[OR/RR_upper − ln OR/RR_lower]) ÷ 3.92. Then, we summarized the overall OR and CI by using general variance-based methodsI2 was used to evaluate the extent of heterogeneity derived from study differences rather than chance.44 The smaller value I2 suggested less obvious heterogeneity. We used the random-effects model to calculate the summary OR and its 95% CI45 with suspecting heterogeneity. We evaluated the impact of the changes on pooled ORs by study design, cancer subtypes, geographical location, source of controls, research quality, and some adjusted confounders such as alcohol and body mass index as prior hypotheses to explain heterogeneity through subgroup analyses and meta-regression analyses. Sensitivity analyses were conducted by removing 1 study from all studies to evaluate the impact on the pooled ORs and heterogeneity. We can, therefore, evaluate whether the results are stable. In an attempt to detect publication bias, we visually examined asymmetry in a funnel plot. We conducted Begg and Egger test to assess whether there is an obvious publication. We considered the funnel plot to be asymmetrical if the intercept of the regression line deviated from zero with P < 0.10. If the test suggests an obvious publication bias, we would conduct the trim and fill analysis to further verify.The value of 46 STATA  was used to conduct the Egger and Begg regression asymmetry test by using the metabias command.47 We conducted the trim and fill analysis to observe whether the results are stable and evaluate the publication bias.We used the Cochrane Collaboration software  to analyze the extracted data with fixed or random-effects model analysis.48 published in 1987 was replaced by Tuyns et al32 published in 1983, as it shared the same database. The result of De Stefani et al49 published in 2003 was replaced by De Stefani et al40 published in 2005, as the latter expanded the sample size based on the former population.The computerized search yielded 433 references, of which 112 were included after abstract review. Citation search identified another 715 articles. Of the 827 articles that were obtained for full-text review, we excluded 808 articles based on the exclusion criteria. In particular, the result of Tuyns et al43 and 13 case–control studies39 . Overall summary OR using the random-effects model showed a 37%, statistically significant reduction in risk of esophageal cancer associated with citrus fruits intake . The subgroup of case–control studies  and the subgroup of cohort studies  showed a respective 46% and 24% statistically significant reduction in risk of esophageal cancer associated with citrus fruits intake ; however, the result of Begg test suggested an obvious publication bias (P = 0.046). And the outcome of trim and fill analysis demonstrated that there was no publication bias.No publication bias was observed in the selected studies. Visualization of Begg funnel plot was symmetrical Figure . Formal P = 0). The result is supported by the strengths of our review, which includes a systematic literature search, strict selection criteria, comprehensive data abstraction, and rigorous statistical analysis. Additionally, the results of similar reviews about the association between citrus fruits and other cancers29 are encouraging.The overall summary OR in our study presents an inverse association between citrus fruits and esophageal cancer , 0.76 (CI 0.62–0.93) for the case–control study and cohort study, respectively. The discrepancies between study results can be explained by recall and selection biases in case–control studies and by imprecise dietary measurements and limited variability of dietary intake in cohort studies.60 To further explore the source of heterogeneity, we conducted subgroups analyses and meta-regression analyses by many factors such as cancer subtypes, geographical location, source of controls, research quality, and adjusted confounders.We found heterogeneity among the included studies, which decreases the quality of evidence to very low quality.21 vary widely, so exploring the impacts of citrus intake on these cancers is essential. Four of the included studies did not describe the specific cancer subtypes or included both subtypes. Table 10 and the summary OR. Repeat meta-analysis of a new model excluded the study10 from all 19 selected articles was conducted.61 The level of heterogeneity decreased from high (I2 = 52%) to low (I2 = 0).The pathogenesis and risk factors for different types of esophageal cancer62 Dietary antioxidants are emerging as potentially modifiable risk factors for EAC. High intake of beta-carotene may be associated with decreased risk of dysplastic Barrett esophagus, which is regarded as the precursor of EAC.63 Some studies65 showed that carotenoids may be responsible nutritional factors  in the development of different malignant diseases including esophageal cancer.67 Carotenoids may intervene in cancer-related molecular pathways and the expression proteins involved in cell proliferation, differentiation, apoptosis and angiogenesis, carcinogen detoxification, DNA damage, and repair.68 A related study indicates69 that a high intake of vitamin C is associated with a reduced risk of EAC and reflux esophagitis. Antioxidants may also play a role in the pathogenesis of reflux esophagitis and EAC and may be more important in terms of progression rather than initiation of the disease process69; however, low intake of vitamin C and E correlates significantly with the development of SCC as well as EAC in males.71 Regarding the mechanism, researchers think that vitamin C could enhance the EGCG- and TF3-induced apoptosis in SPC-A-1 and Eca-109 cells via MAPK pathways.72 Additionally, folate and other dietary methyl group factors are implicated in the etiology of EAC and its precursors. Folate is implicated in carcinogenesis via effects on DNA synthesis, repair, and methylation.22 Some studies indicate that flavanone intake is inversely associated with SCC risk and may account for the protective effect of fruit, especially citrus fruits, on esophageal cancer.74 Because citrus fruits account for 90% of flavanone intake, the findings of Rossi et al73 suggest that flavanones may play a role in the protective effect of citrus fruits on esophageal cancer. Therefore, the basic research of mechanisms flavanones protect against esophageal cancer are worth studying. Although the results are exciting, we have to taken in account the interaction between medicines and fruits. The research by Bailey et al75 suggested that there exist adverse reactions when grape is combined with some drugs.Citrus fruits include many bioactive components.Our review demonstrates that citrus fruit intake could reduce the incidence of esophageal cancer by 37% based on published results of epidemiologic studies. The trends are consistent between case–control studies and cohort studies; however, considering the drawbacks mentioned above, our conclusions should be taken cautiously. There are no relevant studies that provide explicit evidence for the inconsistency between SCC and EAC. The low quantity of EAC cases and the limitations of meta-analysis are responsible for the results. Therefore, larger studies with rigorous and prospective methodology should be considered to validate the association between citrus fruits and esophageal cancer. It is still unknown which components in citrus fruits have an effect on esophageal cancer prevention. Our conclusion may encourage researchers to further explore the protective elements and potential mechanisms, which may contribute to reducing the esophageal cancer risk. We hope further research will explore this issue.


We aim to quantify the effects of nut intake on risk of stroke by a dose-response meta-analysis with a random-effects model. Two databases (PubMed and Emabse) were searched for prospective cohort studies regarding nut intake and stroke risk. Studies were included if they fulfilled the predefined criteria. Eleven articles encompassing fourteen cohort studies were included in final analysis. The pooled relative risk (RR) of stroke for the highest versus (vs.) lowest category of nut intake was 0.88 . The power to detect a RR of 0.88 for the highest versus vs. lowest category of nut intake was 86.2%. In multiple subset analyses by gender, location, and stroke subtype, the inverse association was only found in women  and Asia . In the dose-response meta-analysis, evidence for a nonlinear association between nut intake and stroke risk was observed and a RR of 0.86 was conferred for 12 g/day. Based on the Grading of Recommendations Assessment, Development, and Evaluation (GRADE) system, the quality of evidence was moderate. In conclusions, finding from current meta-analysis of fourteen cohort studies indicates that nut intake may be related to decreased risk of stroke. The detail of search strategy was shown in PubMed  assessing the relationship between nut consumption and risk of stroke; (2) providing estimates of the relative risk and corresponding confidence interval, or other information sufficient to calculate them. If more than one report from the same study were published, the one with the longest years of follow-up were included in the present analysis.The data were extracted by one reviewer and examined independently by another two reviewers. Any discrepancies were resolved by discussion. The available data from each included studies included first author, country where the study was performed, cohort name, sex, people’ age at enrollment, years of follow-up, number of cases observed, methods for the measurement of nut intake, number of total participants, methods of stroke diagnosed, most-adjusted multivariable RR with corresponding 95% CI for the different categories of nut intake , and confounders controlled in adjusted models.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp accessed Feb 14, 2016) to assessed methodological quality. The Newcastle-Ottawa Scale judges a study on the basis of selection of study groups, comparability of groups and ascertainment of the exposure or outcome of interest for case-control or cohort studies respectively. Due to no unified standards well established, we defined that studies with ≥6 stars were considered as at low risk of bias in our study.We used the 9-star Newcastle-Ottawa Scale  was used to assess the quality of evidence. Evidence from cohort studies begins with a grade of “Low”. The quality of evidence was upgraded for large magnitude of effect, plausible residual confounding that would not reduce the effect size, and a dose response gradient, or downgraded for inconsistency, indirectness, imprecision, and publication biasThe RR with corresponding 95% CI was used as the effect size for all studies and the hazard ratios were directly considered as RRs. This method has been widely used in systematic review and/or meta-analysis2521242526We further quantified dose–response relationships of nut consumption and risk of stroke based on the method described by Greenland and Orsini N3032 statistic322 test range from 0 to 100%. In our study, values of the I2 statistic < 25% are representative of low heterogeneity and those > 75% of high heterogeneity. Correspondingly, the value ranging between 25 and 75% represents moderate heterogeneity.Homogeneity across studies was tested by Q statistic and quantitatively measured by the IAll statistical analyses listed above were done with STATA 12.0 software . P values less than 0.05 were considered statistically significant, except where otherwise specified.et al.Power calculation with the method proposed by Cafri The flow diagram of screened, excluded, and included studies was shown in 68101213141516178681012131415161718810132 = 0%). Seven of the eleven articles addressed the association between nut intake and stroke mortality61214151617182 = 0%).Fourteen cohort studies from eleven articles reported the association between nut intake and stroke risk6810121314151617182 = 0%). The corresponding estimate for males is 0.97(95% CI 0.81–1.16), with moderate statistically heterogeneity .Nine articles provided sufficient data for subgroup analysis681012131415172 = 0%; Asia: P = 0.825, I2 = 0%), but moderate statistically heterogeneity was observed for Europe .Of the fourteen cohort studies from eleven articles, eight were recruited in United States68101213151416182 = 39.2%). Three articles provided data on hemorrhagic stroke8132 = 50.2%).Five articles provided data on ischemic stroke81013172 = 0%). Among the cohort studies with less than ten years of follow-up1516172 = 4.9%).The summary RR was 0.87(95% CI 0.77–0.97) for those with a median follow-up of ≥10 years68121317The influence of single study on overall result was estimated . A bordeThe funnel plot appears to be symmetrical, indicating that there is no evidence for publication bias . Also, tSeven articles were identified in dose-response analysis121314151718The power was 86.2% to detect a RR of 0.88 for highest versus lowest category of nut intake . On the Our meta-analysis of 14 prospective cohort studies from eleven publications provides evidence that nut intake is inversely associated with risk of stroke and a linear association between nut intake and stroke risk did not exist.Nut contains a variety of nutrients with anti-inflammatory, antioxidant, and anticarcinogenetic properties, involving the unsaturated fatty acids, folate, niacin, vitamin E, and vitamin B-6, dietary fiber, phytoestrogens, and micronutrients 35In subgroup analyses by gender and geographic area, significantly inverse association between nut consumption and stroke risk was only observed for female and Asia subsets. Common explanations for these discrepancies are (1) different dietary habits and genetic backgrounds were shared, and (2) the chance results were identified, as a small number of studies were included in the subgroup analyses.Hemorrhagic and ischemic strokes vary in pathogenesis and prognosis. With regard to the incidence, ischemic strokes are responsible for about 87% of all cases. However, hemorrhagic strokes account for about 40% of all stroke deaths, although hemorrhagic strokes are less common. Moreover, some studies showed people bear different risk from the same risk factorUp to now, it remains unclear that whether different types of nuts have different effects on stroke. To our knowledge, there only two articles performed separate analyses of the consumption of peanuts and tree nuts with stroke risk12The major concern about safety of nut consumption is possible weight gain3539Several meta-analyses of the relationship between nut intake and stroke risk have been published. The discrepancies between previous meta-analyses and ours should be noted . A majorLike others of similar design, our met-analysis has some limitation. First, the effect of residual confounders on the association cannot be elucidated utterly, although we used the most-adjusted RRs. Second, some included studies reported data about total nuts and peanut butter, and thus an underestimation of the risk with the true amount of nut consumed may be resulted out. Third, the issue whether the relationship between nut intake and stroke risk varies by the prepared method was not addressed, as none of included studies provided information. Fourth, potential publication bias is an intimidation to robustness of our findings, although there was no evidence for publication bias.Our meta-analysis of published prospective cohort studies supports nut intake is associated with a reduced risk of stroke. The risk of stroke with nut intake seems to be lower in Asian. The reduced risk was pronounced at the level of 12g/gay. The quality of evidence was moderate. This finding supports daily nut intake is an important part of healthy lifestyles and could be recommended in the prevention of stroke.How to cite this article: Shao, C. et al. Nut intake and stroke risk: A dose-response meta-analysis of prospective cohort studies. Sci. Rep.6, 30394; doi: 10.1038/srep30394 (2016).


WNK4) gene polymorphisms and hypertension has been widely investigated, However, the studies yielded contradictory results. To evaluate these inconclusive findings comprehensively, we therefore performed a meta-analysis. Ten articles encompassing 16 independent case-control studies with 6089 hypertensive cases and 4881 normotensive controls were selected for this meta-analysis. Four WNK4 gene polymorphisms were identified . The results showed statistically significant associations of G1155942T polymorphism  and C6749T polymorphism  with hypertension risk. However, neither C1155547T nor G1156666A was associated significantly with hypertension susceptibility. In conclusion, this meta-analysis suggested that WNK4 G1155942T and C6749T gene polymorphisms may contribute to the susceptibility and development of hypertension. Further well-designed studies with larger sample size are required to elucidate the association of WNK4 gene multiple polymorphisms with hypertension risk.The relationship between with-no-lysine [K] kinase 4 ( Hypertension is a worldwide public health problem; its overall prevalence is approximately 30–45% in the general population with a steep increase with agingBP is regulated by a complicated network, including vascular, renal, neuronal and endocrine mechanisms and is also influenced by environmental factors, genetic factors, and multifactorial interactions78WNK1 and WNK4 are considered to be related with pseudohypoaldosteronism type II (PHA II), an autosomal dominant disease featuring hypertension, hyperkalemia, hyperchloremia and metablic acidosis13WNK1 and WNK4 were involved in regulation of diverse ion transporters and channels of the distal nephron151617WNK1 and WNK4 led to deregulated renal sodium absorption, potassium secretion, thereby giving rise to the PHA II phenotypeWNK1 are large deletions of the first intron that lead to the increased expression of wildtype WNK1, mutations in the WNK4 gene are missense mutation and cluster within the highly conservative coding sequence outside the kinase domain11Up to now, a serial of previously unrecognized kinases interacting each other in the distal nephron have been identified as playing important roles in sodium, potassium, and BP regulation. WNKs (with-no-lysine [K] kinases) are a novel family of serine-threonine protein kinases. They are different from other usual protein kinases in subdomain II of WNKs that lacks the common conserved catalytic lysine residue that is crucial for binding to ATP and protein phosphorylation11WNKs gene variation and hypertension susceptibility, since Wilson FH et al. reported their potential relevance each other by studying a PHA II kindredet al. identified a significant association of WNK1 AluYb8 insertion with BP variation in the Estonian HYPEST cohort study and confirmed this result by a meta-analysis of three independent European samplesWNK1 common and rare single nucleotide polymorphisms (SNPs) and haplotypes on BP have been found in both general population and hypertensives212223WNK4 polymorphisms with hypertension susceptibility is ambiguous on account of the multiple potentially related gene loci and inconsistent results have been reported in different studies252627WNK4 polymorphisms and hypertension risk as we know. Therefore, in order to estimate the relationship between WNK4 gene polymorphisms and hypertension, especially G1155942T, G1156666A, T1155547C, and C6749T, we performed this meta-analysis.Subsequent numerous epidemiological studies were conducted to evaluate the relationship between et al.et al.et al.WNK4 gene polymorphisms and hypertension including 6089 cases and 4881 controls. Main characteristics of these included studies are shown in et al.WNK4 SNPs and hypertension risk are summarized in According to the inclusion criteria, 10 eligible studies252628293031323334P = 0.11, P = 0.35) and homozygous model . Because of the absence of wild homozygote (CC) in both case and control group, the study by Lu M et al.Pheterogeneity < 0.1, I2>50%), while fixed effects model was used in homozygous model.In total, three studies2531WNK4 G1156666A polymorphism with hypertension. Heterogeneity between studies was significant in allele contrast and dominant model , and thus we used a random-effects model for pooled analysis. One study by Erlich et al.WNK4 G1156666A polymorphism and hypertension risk .There were four articles252632WNK4 G1155942T polymorphism with hypertension, five studies with 2260 cases and 1567 controls were pooled in this meta-analysis25283032et al.P = 0.01) and dominant genetic model . However, no significant relation was found in homozygous genetic model   .WNK4 C6749T polymorphism with hypertension was found to be significant in allele contrast , dominant genetic model  and homozygous genetic model  , a fixed-effects model was used.The association of  = 0.02) . As the P-value of Egger's test confirmed the existence of publication bias regarding the WNK4 C1155547T polymorphism and hypertension under allele contrast (P = 0.02). No statistical significance of publication bias was observed by Egger's test to G1156666A (P = 0.50), G1155942T (P = 0.18) and C6749T (P = 0.75) polymorphisms under allele contrast.Both the Begg's test and Egger's regression test were performed to assess publication bias. The shapes of the funnel plots do not show obvious evidence of asymmetry . HoweverThe sensitivity analyses were performed by sequential omitting individual studies to examine the influence of individual studies on the pooled ORs. The pooled ORs and corresponding 95% CIs were not significantly altered after excluding individual study in sequence see  online, WNK4 SNPs and hypertension were performed according to the year of publication. Regarding to C1155547T, as studies published in order, the statistical results had a trend to close to significant gradually and slightly. However, for G1156666A, G1155942T and C6749T polymorphisms, no obvious evolution rules were observed as records accumulated by year , the epithelia sodium channel (ENaC), the K+ channel (ROMK) and the paracellular Cl− pathways principally15163738WNK4 may cause increased NCC expression in the DCT and increase paracellular Cl− permeability in the distal tubule and reduce the surface expression of ROMK channels1940+ retention in the DCT contributing to development of hypertension. Consistently, the results of this meta-analysis confirmed the association of WNK4 gene variation with the susceptibility of hypertension.Human et al. firstly reported that mutations of WNK1 and WNK4 genes might be associated with PHA II, a rare autosomal dominant disease with the main performance of hypertensionWNK1 and WNK4 gene mutations and hypertension or blood pressure regulation. Studies by Putku et al.et al.WNK1 gene were related with BP variation and replicated the association in their meta-analysis as well. However, no meta-analysis examining the association of WNK4 gene mutations with the risk of hypertension existed. This study emerged under the circumstances in order to elucidate the role of WNK4 gene polymorphisms in development of hypertension. According to the limited quantities of relevant eligible studies, we only selected four polymorphisms in WNK4 gene  with at least 3 available relevant studies for further analyses. The mutations of WNK4 gene G1662A, K1169E, Pro556Thr and Q565E were also suggested to be associated with blood pressure4243Wilson FH WNK4 gene polymorphisms and hypertension as reliable as possible, some limitations of our study should be concerned. First, the included eligible studies without of large enough sample size limited the statistical power to some extent. Moreover, the quantities of each studied gene polymorphism were quite small and no sufficient data could be included for further analysis when we attempted to perform subgroup analyses. More studies with lager sample size are needed in the future to better interpret our results. Second, although the comprehensive literature search was performed in both English and Chinese database without language restriction, however, the P-value of Egger's test indicated the existence of publication bias for C1155547T polymorphism in association with hypertension. Third, the pathogenesis of hypertension is considered to be quite complicated involving gene-gene and gene-environment interactions. In this meta-analysis we tried to collect all the potentially relevant information such as BMI, gender distribution of the studied population, age and the level of serum lipid, however, we failed to assess the effect of gene-gene and gene-environment interactions. It has been reported that WNK kinases are upstream activators of ste20-related pralinealanine-rich kinase (SPAK) and ORS1 and the WNK-SPAK/OSR1 signaling pathway is very important in regulation of ion cotransporters and blood pressureSTK39 gene encoding SPAK protein has been confirmed to be associated with hypertension by a meta-analysis recentlyCaMK4) and G-protein-coupled receptor kinases (GRKs) are also demonstrated to related to regulation of blood pressure47Though we tried our best to control the potential bias from statistical aspect and estimated the association between WNK4 gene may increase the susceptibility of hypertension. Whereas, biological evidence was absent to support the association between WNK4 C115547T and G1156666A polymorphisms and risk of hypertension. Hypertensive disease is one of the major causes of mortality and global burden of disease, timely recognition and effective intervention of hypertension can help control the hospital mortality of cardiovascular disease49In conclusion, the current meta-analysis suggested that the G1155942T and C6749T polymorphisms of WNK”, “variant”, “polymorphism”, “blood pressure”, “high blood pressure” and “hypertension” were used. We also manually searched the reference lists of the included studies to find additional eligible studies. The literature search was finally performed on March 20, 2014.A comprehensive literature search for relevant articles without language restrictions was performed through the following electronic database: PubMed, MEDLINE, Web of Science, China National Knowledge Infrastructure (CNKI), Wanfang Data and China Biology Medicine (CBM). To identify all possible studies as complete as possible, various keywords as “with-no-lysine kinase”, “The studies were included in this meta-analysis if they comply with the following inclusion criteria: 1) evaluated the WNK4  polymorphisms and hypertension; 2) used case-control or cohort design; 3) provided sufficient data of sample size, genotype distribution and allele frequencies or other information such as ORs with 95% CIs for statistical analysis; 4) diagnosis of hypertension patients based on the criteria of systolic blood pressure ≥ 140 mmHg or diastolic blood pressure ≥ 90 mmHg. If multi-studies reported the same or reduplicative data, we selected the latest published or the one with the largest sample size.Two authors (X. G. Guo and J Ding) extracted the following information from each qualified studies independently: first author's name, publication year, ethnicity of study population, study design, genotyping methods, source of the controls, sample size of cases and controls, the distribution of genotypes in cases and controls, and diagnostic criteria for hypertensive cases and normotensive controls. We also collected the baseline information of each study population, such as age, gender, body mass index (BMI), mean blood pressure of cases and controls, and serum concentration of total cholesterol (TC) and triglyceride (TG). If the inconsistent evaluations are encountered, the two authors reached a consensus by discussion or intervention of a third author.WNK4 polymorphisms and hypertension. The heterogeneity among different studies was assessed by chi-square-based Q-testsI2 was used to quantify the effect of heterogeneity. If the P-value of the Q-test >0.10 or I2 <50%, we chose a fixed-effects model by Mantel-Haenszel method for the absence of significant heterogeneity. Otherwise, a random-effects model with the method of Dersimonian & Laird was used52ORs and corresponding 95% CIs were used to evaluate the possible association between We also performed sensitivity analysis by omitting an individual study each time to check whether any of these estimates can bias the overall estimate. Furthermore, cumulative meta-analysis was performed to estimate the influence on the subsequent studies by the first published study, and identified the evolution of pooled estimates as time goes byP < 0.05 was considered the existence of statistically significant publication bias. The HWE of controls was calculated using Pearson χ2 -test. The genotypes and allele frequencies of controls were considered in HWE if P > 0.05. All statistical analyses were performed using STATA 11.0 .Finally, publication bias was assessed by both Begg's testConceived and designed the experiments: G.X.G., D.J. Performed the experiments: X.T.M., J.W.Q., Y.X. Analyzed the data: S.Y.P., Z.F.R., Contributed reagents/materials/analysis tools: Z.J.H., Z.L.R., Wrote the paper: D.J., X.H., G.X.G.Supporting Information


We aimed to examine the association between chocolate intake and the risk of incident heart failure in a UK general population. We conducted a systematic review and meta-analysis to quantify this association.We used data from a prospective population-based study, the European Prospective Investigation into Cancer (EPIC)-Norfolk cohort. Chocolate intake was quantified based on a food frequency questionnaire obtained at baseline (1993–1997) and incident heart failure was ascertained up to March 2009. We supplemented the primary data with a systematic review and meta-analysis of studies which evaluated risk of incident heart failure with chocolate consumption. A total of 20,922 participants  were included of whom 1101 developed heart failure during the follow up . After adjusting for lifestyle and dietary factors, we found 19% relative reduction in heart failure incidence in the top (up to 100 g/d) compared to the bottom quintile of chocolate consumption (HR 0.81 95%CI 0.66–0.98) but the results were no longer significant after controlling for comorbidities (HR 0.87 95%CI 0.71–1.06). Additional adjustment for potential mediators did not attenuate the results further. We identified five relevant studies including the current study . The pooled results showed non-significant 19% relative risk reduction of heart failure incidence with higher chocolate consumption (HR 0.81 95%CI 0.66–1.01).Our results suggest that higher chocolate intake is not associated with subsequent incident heart failure. •Whether chocolate consumption increases or decreases the risk of heart failure remains controversial.•Our results suggest that higher chocolate intake is not associated with subsequent incident heart failure.•Our meta-analysis found consistent findings across all studies that chocolate consumption does not increase the risk of incident heart failure.•Evidence suggests that chocolate consumption does not increase risk of heart failure at the general population level. Cocoa, one of the main ingredients in plain chocolate, is an important dietary source of flavonoids which are believed to have cardiovascular benefits. Both observational and trial evidence have shown that chocolate consumption is associated with a reduction in blood pressure Existing observational studies evaluating the risk of heart failure with chocolate consumption show conflicting results and are not directly applicable to the general population. They were conducted in highly selective populations such as women who underwent mammography In this paper, we report results from a large prospective UK population study, the European Prospective Investigation into Cancer-Norfolk (EPIC-Norfolk), and quantify the effect of chocolate consumption on heart failure incidence by conducting a systematic review and meta-analysis, including the current study.The study methods of the European Prospective Investigation into Cancer (EPIC)-Norfolk have been previously described in detail The data collection methods are described in A one year recall of chocolate consumption was estimated using three questions from the FFQ and these questions were consumption of: “Chocolates singles or squares” (average portion size of 8 g), “Chocolate snack bars, e.g. Mars, Crunchie” (average portion size of 50 g) and “Cocoa, hot chocolate (cup)” (average portion size of 12 g powder weight). The amount of chocolate product eaten (grams/day) was derived from multiplying the frequency categories by the portion size. The sum of the weights of these food items consumed, rather than their flavonoid or cocoa content, formed the measure of exposure. Chocolate intake ranged from 0 to 348 g/d. We excluded participants with chocolate intake greater than 100 g/d (n = 68), since generalization of the results to such high quantities seemed unjustified. Equally, errors (i.e. outliers) relating to participant's poor comprehension of FFQs in general, or specifically the questions on chocolate products, were hereby minimized.East Norfolk COmmission REcord) database. All participants were flagged for death certification at the UK Office of National Statistics, ascertaining vital status for the entire cohort. Incident heart failure cases were ascertained by using death certificate data and hospital record linkage using the “International Classification of Disease-10” (ICD10) code I50 with virtually complete follow up Hospital admissions for participants were identified using the participant's unique National Health Service number and linking this to the ENCORE . We categorized chocolate consumption into quintiles. The participants in the first quintile consumed no chocolate. Baseline characteristics were compared between the quintile groups of chocolate consumption using one-way analysis of variance (ANOVA) for continuous variables and chi-squared test for categorical variables. We excluded participants with prevalent baseline heart failure , physical activity (active vs non-active), energy intake (<median energy vs ≥median energy), prior myocardial infarction, prior diabetes and prior arrhythmia. We formally tested for interaction by comparing the model with and without interaction term using the P-value of the likelihood ratio test (LRT). The interaction terms were presented graphically.The five different levels of adjustment for confounders are described in We also conducted a sensitivity analysis excluding participants who had incident heart failure within the first 3 years of the follow-up.2 where values greater than 50% were considered to have substantial heterogeneity We aimed to identify studies that reported on the association between chocolate consumption and incident heart failure. We searched PubMed and EMBASE from inception until December 2013 using the terms (cocoa OR chocolate) AND (heart failure OR cardiac failure OR congestive cardiac failure OR left ventricular dysfunction), with no language restrictions and we checked bibliographies of included articles. We set up an automated search on PubMed for updates (most recent April 2015) on newly published studies with terms such as ‘chocolate’ and ‘heart or cardiac or cardiovascular’. We also performed a search of ESC 365 for unpublished studies. Two reviewers independently screened abstracts and titles, and then obtained full-text versions of potentially relevant studies to confirm eligibility. Studies were included if they evaluated chocolate intake and incident heart failure. Data extraction of included studies was performed by CSK and this was checked by YKL. We pooled data using the inverse variance method and random effects model in RevMan 5.2 software . For these comparisons, we considered subgroups of women, men, both genders combined and all studies irrespective of gender. We used the multivariable adjusted measures of association  for the highest category of chocolate consumption versus the lowest category of consumption. Heterogeneity was estimated using IAfter excluding participants with prevalent cancer and heart failure at the baseline, a total of 20,922 participants, 9738 men and 11,184 women were included in the study . The cohThe characteristics of participants according to quintiles of daily chocolate consumption are shown in Secondary analyses were conducted considering subgroups of participants . The higOur search yielded 50 potentially relevant articles and 10 conference presentations. From these results, 4 studies met the inclusion criteria . DetailsAll four studies used questionnaires to ascertain chocolate consumption, but there was heterogeneity in the way chocolate consumption was analyzed as one study considered frequency of chocolate consumption  2 = 46%)  . ConsideIn this large prospective population based cohort study, systematic review and meta-analysis, we found the association between chocolate consumption and reduced subsequent risk of heart failure but this was not sufficiently robust after adjusting for potential confounders. This applied to men and women as well as those with or without baseline heart failure risk factors such as myocardial infarction, diabetes and arrhythmias. Considering these results in the context of existing studies, there appears to be consistent findings across all studies that chocolate consumption does not increase the risk of incident heart failure. We observed modest heterogeneity across studies and no significant association.The EPIC-Norfolk cohort represents an apparently healthy middle and older age community dwelling population and thus the current study addresses previous limitations such as data from specific patient populations, or focused on a particular sex with certain characteristics and risk profile Previous mechanistic studies have suggested that chocolate consumption could lower the risk of incident heart failure through several pathways. Cocoa is rich in flavonoids which are believed to be responsible for the cardioprotective properties of chocolate. Flammer et al. randomized participants with heart failure to flavanol-rich chocolate or placebo and found that chocolate led to peripheral vasodilatation, improvement in endothelial function and inhibition of platelet function There is also evidence to suggest that chocolate has anti-inflammatory It should be noted that the chocolate consumption in the EPIC-Norfolk was skewed (the median consumption (IQR) was 4.6 (0.56–12)) and we also truncated chocolate consumption at max of 100 g/d to exclude outliers. We thus further examined the effect of incremental increase in chocolate consumption and incident heart failure and there was no evidence for harm (data not shown). We have recently reported that chocolate consumption was associated with favorable cardiovascular outcomes While all studies used food frequency questionnaires to estimate chocolate consumption, we note that chocolate intake had no consistent definition across the studies. Most of the studies considered chocolate from chocolate bars while the study by Lewis et al. considered only solid chocolate and chocolate containing beverages We observed that among the highest consumers of chocolate who had energy intake below the median had a lower risk of heart failure. It is unclear why this is. One possible reason for this observation may be that low energy intake may reflect individuals who are of younger age and younger age is associated with lower heart failure risk.While we found that chocolate intake may possibly be associated with lower likelihood of heart failure, the association was not statistically significant after adjustments for potential confounders and mediators. There could still be a significant association but this study may not have had sufficient sample size and power to detect this. This is supported by the meta-analysis which pooled existing studies which all seem to suggest a lower risk in heart failure with higher chocolate consumption. However, the positive associations from published literature may also stem from publication bias whereby studies that show significant associations are more likely to be published whereas those with null or non-significant findings remain unpublished. More studies with high quality methods and adequately powered sample size are needed to determine if there are benefits of chocolate in terms of heart failure risk.Our study has several strengths. The current study is based on a large population-based cohort thus is applicable to the middle and older aged general population. The characteristics of the EPIC-Norfolk participants were comparable to other UK representative samples except with lower prevalence of current smokers Our study has some limitations. The FFQ has limitations associated with it, such as assumptions about portion size, possible errors in completing the questionnaire due to memory and extended recall time of 1 year. Some participants may under-report especially with chocolate consumption. This could mean that the high chocolate consumers are people with more ‘reliable chocolate intakes’ and possibly with other positive behaviors or characteristics which lower risk of a cardiac outcome. We were unable to differentiate plain, milk and dark chocolate in this study and therefore unable to determine the effect of the different types of chocolate. Mild cases of heart failure might have been missed at baseline. However, sensitivity analysis excluding those diagnosed with heart failure within the first 3 years of follow up did not alter the results. Studies have shown that many patients with left ventricular systolic dysfunction are asymptomatic, undetected and untreated It is worth noting that while chocolate consumption may be associated with potential cardiovascular benefit with regard to incidence of stroke and ischemic cardiovascular disease; the different pathophysiological process which underpins the development and progression of heart failure is very different from ischemic pathology. The initial inverse association between higher chocolate consumption and incident heart failure was attenuated by adjustment of confounders and potential mediators. Nevertheless, we observed an unlikely increased harm from higher chocolate consumption with regard to the risk of developing subsequent heart failure in this general population.PKM, AAW and CSK conceptualized and designed the study. RNL was responsible for data management and CSK analyzed the data and drafted the first draft of the manuscript. CSK and YKL led the systematic review with input from AAW and PKM in study protocol development. KTK and NJW are the Principal Investigators of EPIC-Norfolk cohort. All authors contributed to the study design and writing of the paper. PKM is the guarantor.The EPIC-Norfolk study was supported by grants from the Medical Research Council and Cancer Research UK.The authors report no relationships that could be construed as a conflict of interest.


I2 = 74.2%) but not the mortality associated with it . Specifically, frequent aspirin use was associated with reduced pancreatic cancer incidence . The summary ORs regarding the incidence of pancreatic cancer and either non-aspirin NSAIDs use  or overall NSAIDs use  were not significant. In conclusion, aspirin use might reduce the incidence of pancreatic cancer; however, this finding should be interpreted with caution because of study heterogeneity.Although there is evidence that non-steroidal anti-inflammatory drugs (NSAIDs) might be able to prevent pancreatic cancer, the findings from epidemiological studies have been inconsistent. In this paper, we conducted a meta-analysis of observational studies to examine this possibility. We searched PubMed and Embase for observational (cohort or case-control) studies examining the consumption of aspirin and other NSAIDs and the incidence of or mortality rates associated with pancreatic cancer. Twelve studies including approximately 258,000 participants in total were analysed. The administration of aspirin significantly reduced the incidence of pancreatic cancer  = 0.77; 95% confidence interval (CI) = 0.62 to 0.96;  Pancreatic cancer is the fourth leading cause of cancer-related death and one of the ten most common types of malignancies in the worldet al.et al.P = 0.069)Cyclooxygenase-2 (COX-2) might play a role in cancer development359131415161718192021222324252627Thus, these previous meta-analyses must be updated to revise the evidence concerning this issue. Moreover, the frequency and duration risks of aspirin use in real-world practice have not been systematically evaluated. Therefore, based on recent evidence, we performed a systematic review and meta-analysis of 12 observational studies to explore the possibility that NSAIDs use reduces the incidence of pancreatic cancer in real-world settings.et al.et al.et al.et al.17Our initial search yielded 2,718 potentially relevant publications, of which 283 duplicates were excluded. We then excluded 2,410 studies that were deemed irrelevant to the meta-analysis based on their titles and abstracts. After reviewing the full texts of the remaining 25 studies, we identified 12 studies for inclusion in the meta-analysis  for the case-control studies and 0.83  for the cohort studies. We conducted a subgroup analyses that considered numerous covariates that might have contributed to the overall heterogeneity .Six case-control and two cohort studies were included in this combined analysis, and a pooled estimate  revealed a decrease in the incidence of pancreatic cancer, with moderate heterogeneity amongst studies  nor low-dose aspirin use was significantly related to pancreatic cancer prevention . Significant declines in the incidence of pancreatic cancer were related to high-frequency  and medium-frequency  aspirin use. Only one studyFurthermore, we examined the dose, frequency and duration risks of aspirin use; these data are presented in P = 0.676; I2 = 0.0%; P = 0.451; I2 = 0.0%; The analysis of 3 (two cohorts and one case-control) studies suggested that non-aspirin NSAIDs use was not associated with a decrease in the incidence of pancreatic cancer ; these studies showed no heterogeneity . No significant heterogeneity was observed between the 2 studies  is inversely related to the risk of pancreatic cancer. In contrast, neither overall NSAIDs use nor non-aspirin NSAIDs use was associated with a reduced risk of pancreatic cancer.et al.et al.et al.The main finding of our meta-analysis contradicts a previous meta-analysis  included. In fact, these patients might have died from causes other than pancreatic cancer, thereby creating bias and obscuring the true incidence of pancreatic cancer. Unfortunately, we did not obtain the data needed to evaluate the effect of duration risk on mortality. However, an individual patient data analysis of 8 RCTs suggested that daily aspirin use over at least 5 years significantly reduces mortality due to several common cancers, including pancreatic cancer (hazard ratio (HR) = 0.25; 95% CIs = 0.07 to 0.92)et al.P = 0.014)1618P = 0.097; I2 = 0.0%). In addition, we did not include 5 of the previously included studies in our analysis of the effect of low-dose aspirin use on the incidence of cancer. Of these 5 studies, one did not meet our inclusion criteria21et al.We also investigated the potential implications of aspirin use dose, frequency and duration. A meta-analysis by Cui Our analysis identified a significant negative relationship between aspirin use and the incidence of pancreatic cancer when the case-control studies were pooled together; however, this finding was not true for the two cohort studies. However, the exposure assessments of these cohort studies were based on self-reports. Moreover, these studies were both based on women, whereas the case-control samples included both men and women. Therefore, their results might be biased, and the limited data from the 2 cohort studies might limit the interpretation. In addition, our stratified analysis based on geographic region did not reveal significant results in America or Europe, which raises interesting questions. In fact, all of the included American studies reported baseline aspirin use values exceeding 40%, which might have reduced the chance of observing small relative differences in aspirin use as a risk factoret al.Several mechanisms might underpin the associations observed in our study. Our meta-analysis did not find that the negative association between aspirin use and the incidence of pancreatic cancer depended on dosage. Individuals consuming high doses of aspirin primarily did so for pain control or anti-inflammation purposesGiven the poor prognosis of pancreatic cancer and the widespread use of aspirin in the general population, successful prevention might have a significant public health effect. Our meta-analysis addresses the beneficial effects of aspirin use; however, physicians should be aware of the optimal dosage and frequency of aspirin use as well as its side effects.Several limitations should be acknowledged. First, significant heterogeneity was observed across the studies included in the current analysis. Although we stratified the data into subgroups based on geographic region, gender, study quality, pattern of aspirin use, and adjusted confounds, considerable heterogeneity remained. This result is not surprising given the discrepancies of each study with regard to their designs; the differing race, age and lifestyle factors of their participants; sample sizes; definitions of drug exposure; follow-up evaluation lengths; and, most importantly, the type and dose of aspirin used as well as its exact administration schedule.Second, most studies adjust for major confounds such as age, sex, BMI, smoking status, and diabetes history using multivariate statistical models. However, the adjusted confounds differed across the studies, which might have affected the overall results. In addition, few studies have adjusted for a family history of pancreatic cancer, which is a significant risk factor for the diseaseA third limitation was the potential misclassification of aspirin use. The consumption levels of the low- and high-dose categories as well as the frequency categories varied across studies; we attempted to minimize this imprecision by pooling the most similar data across the analyses. Nevertheless, this imprecision might have created study heterogeneity in the relevant pooled analysis. Therefore, this result should be considered with caution.Another limitation is that the quantity of studies included in the analysis was not sufficient to evaluate publication bias, although the P-value associated with the Egger’s test was non-significant (P = 0.413). Moreover, we did not find unpublished data to avoid publication bias.Finally, our findings concerning the risks associated with the dose, frequency and duration of aspirin use should be interpreted with caution because few studies were included in the stratified analyses, and most studies lacked information regarding these variables. Of the studies that provided this information, these parameters varied across each trial and therefore would have resulted in invalid statistical analyses for these groups.In summary, the results from the current meta-analysis suggest that aspirin use reduces the incidence of pancreatic cancer but not cancer-specific mortality in the real world. However, these findings should be interpreted with caution because of the considerable heterogeneity amongst the included studies regarding incidence risk and the limited number of included studies concerning mortality risk. Specifically, the negative association between aspirin use and the incidence of pancreatic cancer likely depends on the frequency of this use . Moreover, our meta-analysis did not find any associations between the incidence of pancreatic cancer and either non-aspirin NSAIDs or overall NSAIDs use; the studies included in this analysis had no heterogeneity. However, the small number of studies included in the analyses might limit our interpretations. This study might provide insights that enable a better understanding of the relationship between pancreatic cancer risk and aspirin/NSAIDs use. Future research, especially prospective studies, should be conducted to validate our findings and investigate whether the relationship between aspirin use and the incidence of pancreatic cancer depends on dose or duration of use.This study was conducted using the Cochrane methodology and was reported in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelinesWe searched PubMed and Embase for studies of pancreatic cancer risk and aspirin/other NSAID use published in any language from the inception dates of these databases to June 2015. We searched ClinicalTrials.gov for unpublished studies. The search terms included the generic names of individual drugs, their therapeutic classes, and pancreatic cancer outcome terms . No restStudies were included when they met the following criteria: 1) a cohort or case-control design was used; 2) the exposure of interest was any NSAID use, including aspirin, non-aspirin NSAIDs and overall NSAIDs; 3) the incidence rate, mortality rate, or both of pancreatic cancer was assessed; and 4) an adjusted OR or relative risk (RR) with 95% CIs were provided. Regarding duplicate publications15161716The following information was abstracted from all of the included studies using a standardized data collection form: study name (together with the first author’s name and year of publication), study design, study period, study follow-up, study sample size (including both the numbers of cases and controls or the cohort size), study outcomes, the quality score of each study, the types of NSAIDs and consumption categories employed, the ORs and RRs with corresponding 95% CIs for each category, and the adjusted confounds. We also reviewed the supplementary files of each study and contacted the authors for more detailed information when necessary.We assessed the authenticity and quality of the included studies using the NOSTwo investigators (YP Zhang and YD Wan) independently conducted the literature search, study selection, and data extraction as well as the assessments of quality and risk of bias across studies. Any discrepancies were resolved via consensus.P < 0.1) and by estimating I22I statistic of 25% to 50%, 50% to 75%, and >75% were regarded as low heterogeneity, moderate heterogeneity, and high heterogeneity, respectivelyThe measure of an effect was its associated OR and 95% CIs. Because the absolute risk of pancreatic cancer is low, we generally ignored the distinctions amongst the various measures of relative risk (ORs in case-control studies and RRs in cohort studies); therefore, we reported all of the results as ORs for simplicityTo achieve a clear understanding of the relationship between NSAID use and the incidence of pancreatic cancer, we stratified the data based on drug type  and included two outcomes . The effect estimates were pooled both overall and by study design in the aspirin use analysis. Overall NSAIDs use in this report refers to the studies that reported NSAIDs use but did not provide ratios for the use of aspirin or non-aspirin NSAIDs alone. To analyse the dose, frequency, and duration risks associated with aspirin use, we pooled similar data in each category to conduct the analysis. We defined low-dose aspirin use as 100 mg per day and high-dose aspirin use as 300 mg per day. Low-, medium-, and high-frequency aspirin use were defined as ≤1 day per week, 2–5 days per week, and ≥6 days per week, respectively. A two-tailed P-value of <0.05 was considered significant. All statistical analyses were performed using STATA 12.0 .How to cite this article: Zhang, Y.-P. et al. Aspirin might reduce the incidence of pancreatic cancer: a meta-analysis of observational studies. Sci. Rep.5, 15460; doi: 10.1038/srep15460 (2015).


The relationship between the neutrophil-to-lymphocyte ratio (NLR) and tumours as a prognostic factor has been reported in many studies. In this meta-analysis, we evaluated the prognostic role of the NLR in pancreatic cancer (PC). A systematic search was performed in PubMed and Embase for relevant studies. Data from and characteristics of each study were extracted. A meta-analysis was performed to analyse the prognostic role of the NLR using the hazard ratio (HR) and 95% confidence intervals (95% CI). As a result, a total of 2035 patients in 9 cohorts were included in this meta-analysis. The pooled HR of 1.587  showed that patients with an elevated NLR were expected to have shorter overall survival (OS) after treatment. This meta-analysis suggests that an elevated NLR can be used as a predictor of survival in patients with pancreatic cancer. Pancreatic cancer (PC) results in approximately 227,000 deaths per year4679 studies910111213141516A flow chart of the literature search is shown in 2 = 31.0%, p = 0.170) between the studies, a fixed-effects model was used for the analysis. A pooled HR of 1.587  showed that patients with an elevated NLR were expected to have shorter OS after treatment  the NLR was measured with pretreatment serum-based methods; (2) the association between NLR and OS in patients with PC was reported; (3) hazard ratio (HR) and 95% CI were reported; and (4) all included patients were pathologically diagnosed and did not have any tumours besides pancreatic tumours. The following publications were excluded: (1) studies that did not report HR or 95% CI; (2) letters, reviews, expert opinions, or case reports; and (3) studies that had a sample size less than 50.The following items were collected from each study: first author’s name, year of publication, country of the study population, sample size, sampling time, predominant treatment , cut-off value for the NLR, PC stage and HRs with 95% CI.2 ≤ 50% were considered the values that indicated homogeneity, and a fixed-effects model was subsequently applied. P < 0.10 and I2 > 50% were considered the values that indicated severe heterogeneity, and the pooled HR was subsequently calculated by applying the random-effects model. Begg’s funnel plot and Egger’s linear regression test were used to evaluate publication bias. All statistical analyses were carried out using STATA 12.0 .HRs with 95% CI from each study were extracted to generate a pooled HR. If both univariate and multivariate analyses were reported in the same study, multivariate analysis was chosen for the meta-analysis. We performed a test of heterogeneity using Cochran’s Q test and Higgins I-squared statistic. P ≥ 0.10 and IHow to cite this article: Cheng, H. et al. Prognostic role of the neutrophil-to-lymphocyte ratio in pancreatic cancer: a meta-analysis. Sci. Rep.5, 11026; doi: 10.1038/srep11026 (2015).