The results from 15 articles were pooled. Two authors independently took charge of literature search, article selection and information collection. Overall, circulating apelin concentration was significantly lower in CAD patients (N=1021) than in controls (N=654) , with significant heterogeneity (I2=99.3%) but without publication bias. For the association of APLNR gene rs9943582 polymorphism with CAD (patients/controls: 5975/4717), the mutant T allele was associated with a 5.2% increased risk relative to the wild C allele , without heterogeneity (I2=0.0%) or publication bias. Circulating apelin was increased significantly after surgical treatment for CAD (N=202) , with significant heterogeneity (I2=98.0%). Stratified analyses showed that circulating apelin was significantly reduced in studies with age- and sex-matched patients and controls  and with total sample size ≥125 , relative to studies without matching reports and with total sample size <125. In brief, our results suggested that circulating apelin was a prominent athero-protective marker against the development of CAD.It is well established that apelin-APLNR (apelin receptor) pathway plays a central role in cardiovascular system. In this meta-analysis, we summarized published results on circulating apelin concentration in association with coronary artery disease (CAD),  As released by the World Health Organization, coronary artery disease (CAD) is entitled as the world's biggest killer, contributing to approximately 8.76 million deaths in 2015, up from 5.74 million deaths in 1990 , 4. Howein-vivo and in-vitro studies. For instance, Japp et al in a clinical trial found that acute apelin administration in humans can evoke peripheral and coronary vasodilatation and raise cardiac output [apelin and APLNR genetic polymorphisms in predisposition to CAD risk. What's more, circulating apelin changes after surgical treatment for CAD were also summarized to seek additional supporting evidence.In the present study, we were intended to focus on a promising pathway containing apelin and apelin receptor (APLNR). Apelin, a peptide isolated from bovine stomach extracts, is the endogenous ligand for orphan receptor APLNR . The bioc output . In a moc output . Severalc output –17; howeThe flow diagram of search strategy and study selection for this meta-analysis is presented in Table APLNR gene rs9943582 polymorphism with CAD risk [Table CAD risk –26. As 1CAD risk , 1 studyCAD risk , there wth day after the surgery for CAD.P < 0001), while statistical heterogeneity was significant  , with significant heterogeneity . The cumulative and influential analyses were displayed in P = 0.428) and Egger's test (P = 0.187). In addition, differences of other clinical markers between CAD patients and controls were also tested, as shown in Table I2: 18.5%, 3.2% and 0.0%, respectively). Also, there was no indication of publication bias for all other clinical markers in viewing both Begg's test and Egger's test  and two in APLNR gene (rs948847 and rs9943582) were summarized in association with CAD risk. Considering the small number of studies for polymorphisms rs3761581 (N = 3), rs56204867 (N = 3) and rs948847 (N = 2), with the null association with CAD risk, only genotype distributions of rs9943582 (N = 10) were presented and compared between CAD patients and controls   and a low probability of publication bias . Circulating apelin concentration was increased after coronary operation , but with strong evidence of heterogeneity  (data not shown).For the association of s Figure  (P for BP < 0.001) and Asian  countries, while no significance was detected in populations from Middle Eastern countries. By clinical endpoint, significant reduction in circulating apelin concentration was found in ACS  or CAD  patients when compared with controls, and only marginally significant reduction was found in AMI patients. By apelin source, circulating apelin concentration was reduced significantly when measured in serum  and in plasma . By matching status for age and sex, the reduction magnitude in circulating apelin concentration was stronger in studies with matched patients and controls  than those without matching reports. Finally, median cut-off value based on total sample size of all eligible studies was 125, and circulating apelin concentration was reduced with greater magnitude for studies with total sample size ≥ 125  than those with total sample size < 125 . By obesity status in controls, the reduction in circulating apelin concentration was comparable between studies with normal weight controls (BMI < 25 kg/m2)  and obese controls (BMI ≥ 25 kg/m2) . By hypertension status in controls, the reduction was even stronger in the studies involving normotensive controls  than the other studies .In attempting to seek possible explanations of significant heterogeneity, stratifying studies that assessed circulating apelin concentration between CAD patients and controls by continent, endpoint, apelin source, matching status, sample size, obesity and hypertension were conducted, respectively Table . By contAPLNR gene rs9943582 polymorphism for CAD risk and circulating apelin changes after surgical treatment for CAD. Our meta-analytical results collectively suggested that circulating apelin was a prominent athero-protective marker against the development of CAD. As far as we know, this is so far the first meta-analysis that systematically explores the association of apelin and APLNR, either circulating concentration change or genetic variation, with CAD risk by reviewing the current literature.This meta-analysis summarized the results from 15 published articles with the purpose of assessing circulating apelin concentration between CAD patients and controls, the prediction of via inhibiting the electrical activity of vasopressin-releasing neurons [Previous studies have suggested that circulating apelin was correlated directly with obesity, and in a broader sense, with obesity-related cardiovascular events –31. Apel neurons , 32–34.  neurons , 36. Unfapelin and APLNR genetic polymorphism(s) to CAD risk. Considering the statistical power, only a promoter polymorphism in APLNR gene, rs9943582, was summarized and presented. The mutant allele of this polymorphism was found to confer a marginally increased risk for CAD in overall analysis that was not subject to the disturbance of between-study heterogeneity and publication bias. It is universally believed that CAD is a multifactorial polygenic disease [APLNR gene is extremely small [apelin and APLNR genetic defects is encouraged to aid in unveiling the pathogenesis of CAD.Besides circulating apelin difference, we also examined the susceptibility of  disease , 38, andly small . Althougth day of post-surgery relative to that before the surgery, which can, at least in part, support a causal role of circulating apelin in the aetiology of CAD. Nevertheless, because our results are based on only a small number of studies, it is undoubted that larger studies should be performed to confirm the causal relevance between circulating apelin and CAD in the future.Evidence from circulating apelin changes between CAD patients and controls cannot determine causality, as the changes could be secondary to CAD process itself rather than declaring a causal role. To shed some light on this issue, we additionally summarized the changes of circulating apelin concentration after surgical treatment for CAD, and our results on the basis of 4 studies revealed that circulating apelin concentration was increased remarkably at the 5Several limitations of the present study should be acknowledged. As always, selection bias inevitably arises from a meta-analysis, especially when search scope is restricted to the English-language journals only . HoweverIn conclusion, our meta-analytical results suggested that circulating apelin was a prominent athero-protective marker against the development of CAD. Pending consistent replication and validation, it is anticipated that circulating apelin can be identified as a useful biomarker to allow early detection of individuals prone to the development of CAD, and further it can be proposed as an optional treatment target among high-risk CAD patients. Moreover, further studies are warranted to explore underlying mechanisms.This meta-analysis was based on observational data and it was conducted following the guidelines in the Meta-analysis Of Observational Studies in Epidemiology (MOOSE) statement .Online computational search was confined to four literature platforms - MEDLINE, EMBASE, Web of Knowledge and Google Scholar. Search process was independently accomplished by two authors, Tianbao Chen and Bing Wu, using the same MeSH terms acknowledged by all contributing authors. Search results were merged together, with the removal of duplicates by the EndNote X8 on the Macintosh laptop. In addition, the bibliographies of retrieved articles were also searched to make literature coverage as comprehensive as possible.apelin and APLNR genetic polymorphism(s) with CAD risk, or circulating apelin changes after surgical treatment for CAD. In the meanwhile, population source of each qualified article was inspected to remove duplicate publications from the same research team. In case of any disagreement, a discussion between the two authors was held until reaching a consensus.The full-process of selecting all potential articles was independently accomplished by two authors, Tianbao Chen and Bing Wu, based on two steps. The first step was to remove articles that were not published in the English language, examined apelin in animals or focused on the signalling cascades of apelin-APLNR pathway after reviewing the title and/or abstract. The second step was to access the full text of remaining articles to see whether the results on circulating apelin difference between CAD patients and controls, or the association of apelin and APLNR genes between CAD patients and non-CAD controls, as well as circulating apelin changes after surgical treatment for CAD. Exclusion criteria embraced the publication types in either abstract or poster presentation of international meetings, case report, editorial, letter to the editor, narrative or quantitative reviewer, as well as the investigations on CAD severity or response to drug treatment.Both inclusion and exclusion criteria applied to each potential article. For inclusion criteria, the clinical endpoint should be CAD and the results were available on circulating apelin concentration or genotype distributions of polymorphism(s) in apelin and APLNR genetic polymorphisms with CAD risk or circulating apelin changes after surgical treatment for CAD, as well as the demographic and clinical characteristics such as age, sex, body mass index (BMI), the per cents of hypertension, diabetes mellitus and family CAD history, systolic blood pressure (SBP), diastolic blood pressure (DBP), smoking status and mean levels of circulating triglyceride (TG), total cholesterol (TC), high-density lipoprotein cholesterol (HDLC), low-density lipoprotein cholesterol (LDLC), fasting blood glucose (FBG), fasting insulin, homeostatic model assessment for insulin resistance (HOMA-IR) and creatinine (Cr) in both CAD patients and controls, if appropriate. Information collected by the two authors was computationally compared for accuracy, and any disagreement was checked and solved with consensus.The same two authors, Tianbao Chen and Bing Wu, independently collected the results of interest from each qualified article, including surname of the first author, year of publication, area where study subjects resided in, CAD subtype, diagnostic criteria of both CAD patients and controls, source of controls, apelin assaying method, circulating apelin source, matching status between patients and controls, sample size of each group, circulating apelin concentration between CAD patients and controls or the association of apelin and APLNR genes for CAD risk by using odds ratio (OR) and its 95% CI. The effect-size estimates were computed using the random-effects model. Statistical magnitude of between-study heterogeneity was expressed by the I2, the inconsistency index, which can be interpreted as the per cent of variability due to heterogeneity between studies rather than a sampling error. A larger I2 value indicates a higher probability of heterogeneity, and it is universally accepted that this value over 50% denotes statistically significant heterogeneity. The likely explanations of heterogeneity were assessed by stratified analyses. Cumulative analysis and influential analysis were also presented. Publication bias, a type of bias referring to the decreased probability of studies’ results being published when they are near the null, not statistically significant or otherwise less interesting [Mean differences of demographic and clinical characteristics were compared between patients and controls using the unpaired t test or the Mann-Whitney U test, as appropriate. Circulating apelin changes between CAD patients and controls or after surgical treatment for CAD were expressed by using weighted mean difference (WMD) and its 95% confidence interval (95% CI), and for the prediction of polymorphism(s) in eresting , was graData management and statistical analyses were accomplished with the aid of packages in the STATA/SE software  on the Macintosh laptop.


Its dysregulation has been implicated in a variety of human diseases, including type 2 diabetes mellitus (T2DM). The control of expression of this cytokine is associated with insulin resistance and has a strong genetic influence. In order to understand this relationship, the literature from all case-control studies since 2000 to date was reviewed. The genotypes frequency results presented in ten publications with different ethnicities were compared. The correlation between the TNFA promoter genotypes and the risk of developing T2DM remains controversial due to the many discrepancies between the different studies available. Ethnic differences may play a role in these conflicting results, since the distribution of TNFA promoter polymorphisms is distinctive between individuals of dissimilar racial origin. Hence, although the relationship between T2DM incidence and presence of polymorphisms at position -308 of the TNFA gene is not entirely clear, the results of these studies suggest the need for further investigation.Diabetes mellitus (DM) is considered to be a worldwide epidemic disease and its type 2 form comprises more than 95% of all cases. Tumor necrosis factor-alpha (TNF- Noncommunicable diseases, including type 2 diabetes mellitus (T2DM), are considered the leading cause of death in the world and, in Brazil, constitute the main cause of diseases. T2DM assumes an important role in this context given that it is considered a worldwide epidemic disease; in 2011 it appeared among the ten leading causes of death in the world and it is prospected that number of cases will continue to increase [Diabetes can be defined as a group of metabolic diseases characterized by hyperglycemia resulting from defects in insulin secretion, insulin action, or both. The underlining cause of T2DM can be attributed to a combination of resistance to the action of the hormone, increased production of insulin in a compensatory manner, and inadequate secretory response .β cells [On the other hand, according to a study conducted by Kaprio et al. , the herβ cells .α protein. This molecule is composed of 157 amino acids located on chromosome 6. This cytokine is involved in inflammation, apoptosis, infections, and cancer processes [α and development of insulin resistance.Several reports suggest that the characteristics of each individual, including genetic polymorphisms, may confer differences in the occurrence of diabetes and other complex diseases such as cancer and heart disease . The numrocesses . Two polrocesses . In thisrocesses  demonstrThis systematic review was conducted to provide a comprehensive assessment of the association between the polymorphism in the gene TNFA -308G/A and the presence of T2DM.The literature search was made through the Virtual Health Library (VHL) which uses as bibliographic databases the publications found in sources such as LILACS, MEDLINE, and other information bases such as open educational resources, websites, and scientific events. Any publications with case-control studies, on January 21, 2016, that show the relationship between polymorphisms of TNFA gene -308G/A and susceptibility to type 2 diabetes mellitus (T2DM) were surveyed. The search terms used were as follows: “diabetes” and “TNF” or “tumor necrosis factor”, and “polymorphism”. The items recovered were then selected to be studied focusing on T2DM. There was no language restriction placed in the literature searched.The filters are selected in accordance with the main subjects of the following terms: “type 2 diabetes mellitus”, “tumor necrosis factor alpha”, “single nucleotide polymorphism”, and “genetic polymorphism”. The filter base on the type of study included solely the term “case-control study”. Inclusion criteria were as follows: (1) the study must have been case-control; (2) the study should have assessed type 2 diabetes mellitus and -308G/A polymorphism of TNFA gene, even if associated with other comorbidities; and (3) the study should have provided sufficient data, including the number or frequency of alleles and genotypes. Studies were excluded if (1) there were commentaries, case reports, or non-case-control or meta-analyses studies; (2) they did not report sufficient data; (3) they presented data from other types of diabetes or did not specify what kind patients had; and (4) they showed polymorphism in another coding region of the TNFA gene or (5) if the sample was composed of only DM patients.p value and odds ratio, when calculated.Data from eligible studies were extracted independently in accordance with the inclusion and exclusion criteria. For each study the following characteristics were collected: the authors, the study title, the year of publication, where it was published, the sample size, the TNFA -308G/A genotypes with their corresponding simple frequencies and percentages for the control and the case group, and the study's The literature search found articles published starting from 2000, given that the year the complete mapping of the human genome was made available. Advances in molecular biology technology since this project have been quick and, thus, increasing the number of genetic studies published for the association of certain genetic polymorphisms with various pathologies.In the case of literature search for articles with case studies that related polymorphism position -308 of the TNFA gene with diabetes, the result presented forty-three publications between the years 2003 and 2015, ten of which met the criteria inclusion.Thirteen studies presented results for other types of diabetes; eleven related to type 1 diabetes and two to gestational diabetes. Two others did not specify what type of diabetes had patients in the case group. Two of these studies did not show enough genetic data for analysis. In relation to TNFA gene, seven studies presented polymorphisms in region other than the -308 position and three publications had single nucleotide polymorphism in a gene that was not TNFA. Other studies were excluded due to different methodological designs in their inclusion criteria: two meta-analysis, one being observational and the other prospective and, relative to the sample. Furthermore, there was a publication that presented data of diabetic patient's polymorphisms in both the control and the case group.Finally, a total of ten articles were included in this systematic review. The number and homogeneity of the included studies presented were too limited to allow a meta-analysis.Two studies , has been investigated since diabetics have elevated levels of it circulating. TNF-α has also been implicated as an insulin resistance-causing factor associated with the pathogenesis of T2DM [Gradually, however, emerging studies suggested that several other factors could contribute to the development of T2DM, as the expression of proinflammatory cytokines and other molecules that act in the inflammatory process and that appear to play a critical role in the development of chronic complications of DM . Several of T2DM .α is a proinflammatory cytokine that acts in the regulation of cell proliferation, differentiation, and apoptosis [TNF-poptosis . Among tpoptosis . The prepoptosis . Under tIn this review, several studies with the frequency of genotypes in control and case group were observed for a better understanding of the population genetic profile and a possible association between polymorphism of TNFA gene and T2DM in groups with different ethnicities. The influence of genetic polymorphisms on certain diseases may be found in a population and yet not in another; this may impact the frequencies and distribution of a given polymorphism in distinct population. The difference in influence may be due to racial variantion as well as other factors. . In addiCase-control studies were selected in this review and were made with the people of Spain, Taiwan, Poland, India, Tunisia, Mexico, Hungary, and Morocco; nonetheless a positive association was only found in the Spanish, Hungarian, and Moroccan population.Sefri et al.  in theirSo, when comparing the frequencies of polymorphic genotypes (GA and AA) in the groups of case-control study (described in Risk factors for T2DM are increasingly more prevalent in the population. Risk factors not modifiable or irreversible in nature, such as genetic profiles, refer to the individual characteristics. Even though these factors may not be changed, the identification of their presence in individuals and families may enable health professionals to advise change in the lifestyle of these patients, avoiding early manifestation of T2DM.


Associations between inflammatory gene polymorphisms  and osteosarcoma (OS) risk remain unclear. We conducted a systematic search to retrieve studies that investigated associations between inflammatory gene polymorphisms and OS risk. Nine studies that met the inclusion criteria were finally recruited in this meta-analysis. Overall, there was a significant association between TNF-α 308G/A, IL-10 1082A/G and OS risk, while there was no significant association between TNF-α 238G/A, TNF-β 252A/G and IL-6 174G/C and OS risk. Our subgroup analysis showed a significant association between IL-6 174G/C and IL-10 1082A/G and OS risk in Asians, while no such significant correlation was observed with TNF-α 308G/A, TNF-α 238G/A, TNF-β 252A/G and TGF-β1 29T/C polymorphisms. In Caucasians, there was a significant association between TNF-α 238G/A and the decreased incidence of OS. In conclusion, inflammatory gene polymorphisms play a key role in the occurrence and progression of OS. IL-6 174G/C polymorphism was obviously associated with OS risk in Asians, while TNF-α 238G/A polymorphism seemed to be associated with the decreased susceptibility to OS in Caucasians as Altman and Bland test indicated. Although controversial results were observed between IL-10 1082A/G and OS risk in Asians and Caucasians, it is difficult to make a definite conclusion about the role of IL-10 1082A/G polymorphism in the etiology of OS because our Altman and Bland test showed no good evidence to support a different effect in Asians and Caucasians. Osteosarcoma (OS) is one of the most common primary malignant bone tumors probably due to repeated production of osteoid by malignant cells , 2. OS iThe etiology of OS is complex involving multiple factors, but its actual pathogenesis remains unclear , 2, 7, 8In addition to the above-mentioned risk factors, more studies have verified the importance of genes and gene polymorphisms in the etiology, development and complexity of OS , 4, 8. CInflammation has been increasingly recognized as an important factor in the etiology and development of cancers, for many studies have demonstrated that inflammatory cells play an important role in the tumor microenvironment, where they foster tumor cell proliferation, survival and migration –25 and tWe conducted a systematic online search using ‘PubMed’, ‘EMBASE’, ‘Web of Science’, ‘the Cochrane Library’, ‘Science Direct’, ‘Karger’, ‘Wiley Online Library’ and ‘Springer’ and ‘China WeiPu Library’ to identify case-control studies that investigated the relationship between inflammatory gene polymorphisms and susceptibility to OS. First, the following search terms were used to find out how many inflammatory gene polymorphisms were studied: (‘osteosarcoma’ OR ‘OS’ OR ‘osteosarcoma tumor’) AND (‘polymorphism’ OR ‘single nucleotide polymorphism’ OR ‘SNP’ OR ‘variation’), and then, one-by-one screening was performed by two authors. We found that inflammatory gene polymorphisms TNF-α, TNF-β, TGF-β1, IL-1, IL-6, IL-8, IL-10, IL-12 and IL-16 were studied in case-control studies. Next, we used search terms one-by-one to find out whether there were sufficient articles that reported at least one of the gene polymorphisms identified in our previous search stage. The following search terms were used to find out the eligible studies regarding the association between TNF polymorphisms and OS risk: (‘TNF’ OR ‘Tumor Necrosis Factor) AND (‘osteosarcoma’ OR ‘OS’ OR ‘osteosarcoma tumor’) AND (‘polymorphism’ OR ‘single nucleotide polymorphism’ OR ‘SNP’ OR ‘variation’). The search terms, (‘TGF’ OR ‘Transforming Growth Factor’) AND (‘osteosarcoma’ OR ‘OS’ OR ‘osteosarcoma tumor’) AND (‘polymorphism’ OR ‘single nucleotide polymorphism’ OR ‘SNP’ OR ‘variation’) were used to find the associations between TGF and OS risk. As to IL-6, the following search terms were used: (‘IL 6’ OR ‘Interleukin 6’) AND (‘osteosarcoma’ OR ‘OS’ OR ‘osteosarcoma tumor’) AND (‘polymorphism’ OR ‘single nucleotide polymorphism’ OR ‘SNP’ OR ‘variation’). As to IL-10, the following search terms were used: (‘IL 10’ OR ‘Interleukin 10’) AND (‘osteosarcoma’ OR ‘OS’ OR ‘osteosarcoma tumor’) AND (‘polymorphism’ OR ‘single nucleotide polymorphism’ OR ‘SNP’ OR ‘variation’). And then, these studies were further selected. Furthermore, no language restrictions were applied. Secondary searches of the unpublished literature were conducted by searching the reference lists of the selected studies, reviews and conference reports. Reviews and comments were also examined to further identify eligible studies.The inclusion criteria of our meta-analysis were as follows: (1) case-control studies; (2) evaluation of OS risk and at least one of the identified inflammatory gene polymorphisms were reported; (3) having detailed genotype frequency or numbers of alleles and genotypes between cases and controls or effect size reported or estimated from the articles. The exclusion criteria were: (1) reviews or case reports that were not case-control studies; (2) no available data reported; (3) duplicated reports.Data from the eligible studies were extracted according to the inclusion and exclusion criteria by two authors, and a consensus was reached. For each study, the following data were collected: author list, year of publication, ethnicity, sample size, alleles, genotype of each gene polymorphism, and HWE (Hardy-Weinberg Equilibrium). Incomplete data were also obtained by contacting the respective corresponding author.Odds ratios (OR) and 95% confidence interval (CI) were calculated to evaluate the association between inflammatory gene polymorphisms and OS. An allele contrast model (A vs. G), heterozygote model (AG vs. GG), homozygote model (AA vs. GG), dominant (AA/AG vs. GG), and recessive (AA vs. AG/GG) model were used for TNF-α 308G/A and TNF-α 238G/A polymorphism. An allele contrast model (G vs. A), heterozygote model (AG vs. AA), homozygote model (GG vs. AA), dominant (GG/AG vs. AA), and recessive (GG vs. AG/AA) model were used for TNF-β 252A/G and IL-10 1082A/G polymorphism. An allele contrast model (C vs. T), heterozygote model (TC vs. TT), homozygote model (CC vs. TT), dominant (CC/TC vs. TT), and recessive (CC vs. TC/TT) model were used for TGF-β1 29T/C polymorphism. And the strength of association between IL-6 174G/C polymorphism and OS susceptibility was evaluated by OR and 95% CI according to allele contrast model (C vs. G), heterozygote model (GC vs. GG), homozygote model (CC vs. GG), dominant (GC/CC vs. GG), and recessive (CC vs. GC/GG) model.2 metric value. ORs were pooled by random effect model across the study, which considers both within-study and between-study variation.The assumption of heterogeneity was verified by a chi-squared-based Q statistical test and quantified by ISensitivity analysis was performed to assess the impact of each study on the combined effect of the present meta-analysis, and subgroup analysis was performed according to the ethnicity of the study populations. Altman and Bland test  was condA total of nine studies , 36–40 wP = 0.007; AA vs. AG/GG: OR = 2.90, 95% CI = 1.37–6.13, P = 0.005; IL-10 1082A/G , while no significant association was found between TNF-α 238G/A, TNF-β 252A/G and IL-6 174G/C and OS risk   , IL-6 174G/C mutation might increase the risk of OS in Asian populations . This controversial result might be attributed to the different genetic backgrounds in Asians and Caucasians, as shown by the results of Altman and Bland test , while other factors such as the limited number of included studies or the small number of included cases in individual studies could not be ignored.As shown in Table P = 0.017; GG vs. AA: OR = 1.85, 95% CI = 1.10–3.09, P = 0.020; GG vs. AA/AG: OR = 1.80, 95% CI = 1.10–2.97, P = 0.020, Table P > 0.05)   at position 308, affecting TNF-α gene regulation and associated with altered transcriptional activity in many diseases . It has Another common SNP of TNF-α is TNF-α 238G/A polymorphism. It is recently believed that the A-238 allele at position 238 can significantly reduce the TNF-α production; on the other side, it may be associated with down-regulation of tissue inflammation . InteresTNF-β is a sub-type of the TNF family and believed to be involved in a variety of inflammatory, immune-stimulatory and antiviral responses . MeanwhiThree studies –40 were Interleukins are important members of inflammatory cytokines that have been regarded as key contributors to tumor development with respect to proliferation, apoptosis, angiogenesis and differentiation , 58, 59.Despite the comprehensive analysis on the association between inflammatory gene polymorphisms and OS, our meta-analysis still has some limitations. First, of the nine studies that eventually satisfied the eligibility criteria of the present study, only two studies were included in the pooled analysis of TNF-a 238G/A, TNF-β 252A/G, IL-6 174G/C, and IL-10 1082A/G polymorphisms. Second, the sample size of some studies used in this meta-analysis is still relatively small, which might contribute to the inconsistent results and affect the conclusions drawn. Third, no study was conducted to explore the association between TGF-β1 29T/C polymorphism in Caucasian populations, which may not be sufficient to make a convincing conclusion on all ethnicities and indicate a race-specific effect. Therefore, larger-scale and better-designed studies are necessary to determine the association between inflammatory gene polymorphisms  and OS susceptibility.In summary, this is the first meta-analysis to determinate the association of TNF-α 308G/A, TNF-α 238G/A, TNF-β 252A/G, TGF-β1 29T/C, IL-6 174G/C and IL-10 1082A/G polymorphisms with OS susceptibility. Inflammatory gene polymorphisms play a key role in the occurrence and progression of OS. IL-6 174G/C polymorphism was obviously associated with OS risk in Asian populations, while TNF-α 238G/A polymorphism might be associated with the decreased susceptibility to OS in Caucasians. Although controversial associations were observed between IL-10 1082A/G and OS risk in Asian and Caucasian populations, it is difficult to make a definite conclusion about the role of IL-10 1082A/G polymorphism in the etiology of OS because our Altman and Bland test showed that there was no good evidence to support a different effect in Asians and Caucasians.


ISL1, a master control gene expressed in pericloacal mesenchyme. This study sought to further explore the genetics in a larger set of patients following-up on the most promising genomic regions previously reported. Genotypes of 12 markers obtained from 268 CBE patients of Australian, British, German Italian, Spanish and Swedish origin and 1,354 ethnically matched controls and from 92 CBE case-parent trios from North America were analysed. Only marker rs6874700 at the ISL1 locus showed association (p = 2.22 × 10−08). A meta-analysis of rs6874700 of our previous and present study showed a p value of 9.2 × 10−19. Developmental biology models were used to clarify the location of ISL1 activity in the forming urinary tract. Genetic lineage analysis of Isl1-expressing cells by the lineage tracer mouse model showed Isl1-expressing cells in the urinary tract of mouse embryos at E10.5 and distributed in the bladder at E15.5. Expression of isl1 in zebrafish larvae staged 48 hpf was detected in a small region of the developing pronephros. Our study supports ISL1 as a major susceptibility gene for CBE and as a regulator of urinary tract development.Previously genome-wide association methods in patients with classic bladder exstrophy (CBE) found association with  The bladder exstrophy-epispadias complex  represents the severe end of human congenital anomalies of the kidney and urinary tract (CAKUT), and involves the abdominal wall, pelvis, all of the urinary tract, the genitalia, and occasionally the spine and anus. Birth prevalence rates for the most common defect form, classic bladder exstrophy (CBE), range from 1 in 30,000 to 1 in 50,000, varying among North American ethnic groups with the highest prevalence being observed among Native Americans  and the lowest among Asians 1ISL1 (ISL LIM homeobox 1) gene. Multiple markers in this region showed evidence for association with CBE, including 84 markers with genome-wide significance. A subsequent meta-analysis using data from a previous GWAS by our group of 98 CBE patients and 526 controls of European origin also implicated the 5q11.1-q11.2 locus in CBE risk5To identify susceptibility loci for CBE, we recently performed a genome-wide association study (GWAS) of 110 CBE patients and 1,177 controls of European origin. Here, an association was found with a region of approximately 220 kb on chromosome 5q11.1-q11.2ISL1 region and had previously reached genome-wide significancep values of 10−05 ISL1 during genito-urinary tract development, we performed expression studies in mouse embryos and zebrafish larvae (zfl).The present association study followed-up on the most promising genomic regions based on the results of our previous meta-analysis defined by a total of 12 independent markers. One of these markers resides in the ISL1 locus (p = 2.22 × 10−08). A meta-analysis for the marker rs6874700 of our previous and present study showed a p value of 9.2 × 10−19. The relative risk (RR) (95% CI) for rs6874700 in this meta-analysis was 1.93  , British (n = 40), German (n = 7), Italian (n = 39), Spanish (n = 35), and Swedish (n = 116) origin and 1,354 ethnically matched controls. In controls, there was no evidence for deviation from Hardy-Weinberg equilibrium for any of the 12 SNPs (data not shown). Accordingly we had to exclude 18 of the CBE case-parent trios from North America of European from the transmission disequilibrium test (TDT). Hence the analyzed data set for the TDT comprised 92 case-parent trios from North America of European background. 67–2.23) .Isl1 cKO). Hoxa3 Cre driver mouse line possesses Cre activity in the caudal body regionsIsl1 cKO showed the hypoplasia of GT  expression analysis in the early wt zfl between 22 and 60 hpf (corresponding approximately to 26 somite and pec-fin stage). Using an antisense in situ probe comprising an UTR region of zfl isl1 cDNA  regions  the pronephros containing two glomeruli plus nephrons, is already fully functional. In adult fish the subsequently developing mesonephros will yet become the final kidneys. In mammals however only the hereafter developing metanephros will become the final kidney11 regions , which ws region .ISL1 harboring region with the marker rs6874700 (p = 2.22 × 10−8). A meta-analysis for the marker rs6874700 of our previous and present study showed a p value of 9.2 × 10−19. The relative risk (RR) (95% CI) for rs6874700 in this meta-analysis was 1.93 .Following-up on the most promising genomic regions based on the results of our previous meta-analysis in a total of 360 CBE patients, we were able to replicate our findings on the Isl1 is a transcription factor, containing LIM domains, which has been described as playing essential roles for the regulation of cell lineage commitment in organogenesis. It is essential for development of heart, kidney, neurons and limb1517ISL1 locus are further supported by lineage analysis of Isl1-expressing cells in mouse embryos and mutant mice analysis of Isl1 cKO embryos. Isl1 marks the aPCM, which has been shown to contribute to the embryonic formation of the bladder, rectum, and the external genitaliaIsl1 cell lineages match with the organs reflecting various phenotypes of CBE. In the current study, we found that Isl1 is required for the caudal body formation including bladder and GT. The accurate lineage analyses of Isl1-expressing cells suggest that aPCM cells are specified to the bladder and external genitalia around at E7.5. Hence, the current study revealed the presence of the embryonic-time dependent mesenchymal cell population as sensitive for the current set of genetic mutations. The epithelial-mesenchymal interaction between the epithelia and PCM is necessary to form the major parts of the caudal embryo. During such caudal embryonic development, several signaling pathways such as Bone morphogenetic protein (Bmp) and Sonic Hedgehog (Shh) are activated in the aPCM. On note, Defects of Bmp signals cause the body wall closure defects20Our findings in CBE patients for the isl1 expression close to the proximal region of the developing pronephros for all tested time points. Later during development of zfl, the two nephrons fuse with the caudal end of the intestine, attaching at 96 hpf at the cloacaisl1 could not be detected in the developing more distal cloaca or proctodeum regions in the zfl we analyzed, proximal expression of isl1 at regions of these early structures of mammalian urinary tract equivalents in zebrafish provides additional evidence for ISL1 as a regulator for urinary tract development across species.Using WISH analysis in early developmental stages of zfl (22–60 hpf), we were able to detect p values which were not considered worthy of note We were not able to replicate any of the other 11 regions tested in our single-marker analysis. While we cannot exclude, that the inability to find further evidence for association for these 11 regions is based on a lack in statistical power of our present study, it is more likely, that these regions, respectively markers within these regions, had a higher effect in our previous meta-analysis (‘winner’s curse’)  than in our extended replication sample . Alternatively, these regions might have been false positive signals in our previous meta-analysis, which were unrelated to CBE, or only show association with CBE patients of central European origin but not in CBE patient cohorts of different ethnicity. Finally, and most likely, our previous meta-analysis may have failed to identify the possible relevance of other markers and regions, that are actually involved in human CBE formation but were not considered for replication in our follow-up step since they might have obtained p value of 1.14 × 10−10 in our previous meta-analysis the marker rs2303751 in exon 4 of ISL1 reaches a score of 2b  in “RegulomeDB” (http://regulomedb.org) indicating that rs2303751 resides in a functional region. The marker rs2303751 alters 17 DNA binding motifs according to HaploReg v4.1, including alteration of the position-weight matrix for SMAD4, an essential downstream transcriptional regulator of the TGF beta signaling pathwayISL1 is warranted24Within our replicated region of genome-wide significance and with an achieved et al., looking at methylation in low/intermediate grade non-muscle invasive bladder cancer, showed that the ISL1 genes’ promoter-associated island was more frequently methylated in recurrent and progressive high-grade tumors than their non-recurrent counterpartset al., were just recently able to provide certain lines of evidence, that imprinting might play a role in BEEC formation, further studies regarding differences in methylation of the ISL1 region in CBE patients compared to matched controls are also warrantedIn addition, Kitchen ISL1 as a regulator of urinary tract development.Our present association study in CBE patients together with functional studies in mouse embryos and zebrafish larvae suggest This study was approved by each participating center’s institutional ethics committee. All methods were carried out in accordance with relevant guidelines and regulations. All experimental protocols were approved by the institutional committee of the University of Bonn. Furthermore the study was conducted according to Declaration of Helsinki principles. Written informed consent was obtained from all patients, parents/guardians and the normal controls. Demographic information was collected from both, patients and controls, through a structured questionnaire. Experienced physicians trained in the diagnosis of the BEEC personally recruited all BEEC patients included in this study.The study sample comprised 274 Australian n = 36), British (n = 40), German (n = 7), Italian (n = 40), Spanish n = 35), and Swedish (n = 116) CBE patients and 1,365 ethnically matched controls, and 110 case-parent trios from North America of European background. Peripheral venous blood samples, saliva samples or buccal swab samples were obtained from patients and controls for DNA extraction. More details about the recruitment process for patients and population based respectively phenotype unrelated controls can be found elsewhere and in the Supplementary ‘Materials, Methods’, and Swe, British® SNP genotyping assays  and conventional Sanger sequencing. Sequenom data were analyzed using the Spectrodesigner Software package  and allele peaks were analyzed with the Sequenom Typer Analysis software. TaqMan generated data were analyzed using the ABI 7900HT instrument and the SDS software . Sanger sequencing generated data were analyzed using CodonCode Aligner software package .For SNP-marker genotyping, we used the Sequenom MALDI-TOF mass-spectrometer (MassArray system), TaqManStatistical analysis was performed using SAS software (version 9.1). The standard chi-square test was used to test for deviations from Hardy-Weinberg equilibrium. The two-sided Armitage’s trend test was used to compare genotype distributions between cases and controls separately in each case-control subsample. In order to test for association of the risk alleles with CBE in trios from North America, we applied the transmission/disequilibrium test (TDT)All experimental procedures and protocols were approved by the Committees on Animal Research at Wakayama Medical University and the experiments were carried out in accordance with the approved guidelines. Each mouse strain used in this study Isl1flox/flox conditional mutant allele, Hoxa3-IRESCre, Isl1-mER-Cre-mERm and R26R-LacZ indicator strain was described previously78733For analysis of Hoxa3-IRESCre; Isl1 mutant embryos (hereafter described as Isl1 cKO), we mated Hoxa3-IRESCre; Isl1Flox/+ with Isl1Flox/Flox, and subsequently harvested at E13.5. Embryonic specimens were fixed overnight in 4% paraformaldehyde (PFA)/PBS, dehydrated in methanol, and embedded in paraffin. Serial sections (6-μm thick) were prepared for hematoxylin/eosin (H&E) staining.isl1-expression in zfl mRNA WISH was performed in embryos of different ages . We mainly followed a WISH standard protocol using an anti-Dig Ab-dilution of 1:10:000in-situ probe was designed for binding at the 5′ untranslated (UTR) region of zf isl1 using the following forward primer (CATCATctcgagAGAGTGACATCGACCAGCCTGCTTTCC) and the reverse primer (CATCATggatccGAAATTCCCACACAGCTTGTGGC) on cDNA of 1 dpf whole zfl. The resulting amplified PCR product of the probe was then cloned into SK(-)pBluescript® using BamHI and XhoI sites and sequenced. We used the DIG RNA Labeling Kit (SP6/T7)  to transcribe an antisense (T3) and a sense probe (T7) as negative control (not shown) from the above plasmid. Final imaging of the WISH stained larvae fish was performed using an Eclipse upright microscope (Nikon) equipped with a 10x and 40x objectives (NA = 0.25 and 0.65 respectively) and a DS-Vi1 digital colour camera run by NIS-Element software (Nikon).All experimental procedures and protocols were carried out in accordance with the approved national guidelines. Zfl for the presented experiments were maintained and staged as described elsewhere14How to cite this article: Zhang, R. et al. ISL1 is a major susceptibility gene for classic bladder exstrophy and a regulator of urinary tract development. Sci. Rep.7, 42170; doi: 10.1038/srep42170 (2017).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


To the Editor:Nutrients in January 2017 [We have read the letter by Alizadeh regarding our article entitled “Dietary Patterns and Pancreatic Cancer Risk: A Meta-Analysis” as published in First, we are grateful to Alizadeh et al. for alerting us to the methodological limitations in this manuscript. We accept the methodological limitations in the present meta-analysis. Thus, in this reply, we have endeavored to respond to the point highlighted by Alizadeh. In our analyses, 32 studies  were finally included. As Dr Alizadeh says, there are only seven studies ,5,6,7,8 In conclusion, we show in this reply that the issue raised by Alizadeh et al was not ignored. But our findings are useful as an exploratory meta-analysis to confirm the significant associations between diet and pancreatic cancer and add to the existing literature supporting the concept that diet plays an important role in the prevention of pancreatic cancer.We are grateful to Alizadeh for bringing to our attention to the methodological limitations in the conduct of the meta-analysis, and for giving us an opportunity to reassess our work from his perspective.


Stratified analyses by fish type demonstrated that dark meat fish , especially tuna fish  intake was associated with reduced AMD risk. Evidence of a linear association between dose of fish consumption and risk of AMD was demonstrated. The results of this meta-analysis demonstrated that fish consumption can reduce AMD risk. Advanced, well-designed, randomized clinical trials are required in order to validate the conclusions in this study.The association between fish consumption and risk of age-related macular degeneration (AMD) is still unclear. The aim of the current meta-analysis and systematic review was to quantitatively evaluate findings from observational studies on fish consumption and the risk of AMD. Relevant studies were identified by searching electronic databases (Medline and EMBASE) and reviewing the reference lists of relevant articles up to August, 2016. Prospective cohort studies that reported relative risks (RRs) and 95% confidence intervals (CIs) for the link between fish consumption and risk of AMD were included. A total of 4202 cases with 128,988 individuals from eight cohort studies were identified in the current meta-analysis. The meta-analyzed RR was 0.76  when any AMD was considered. Subgroup analyses by AMD stages showed that fish consumption would reduce the risk of both early  and late  AMD. When stratified by the follow-up duration, fish consumption was a protective factor of AMD in both over 10 years ( Age-related macular degeneration (AMD) is now the leading cause of blindness in developed countries. AMD-related choroidal neovascularization (CNV) or geographic atrophy (GA), in the United States, is expected to increase by 50% by 2020 . Effectin-3 fatty acid consumption and reduced risk of late AMD. In a study of an elderly French population, high concentrations of plasma n-3 fatty acids were associated with a decreased risk of late AMD [As oxidative stress is one of the key pathogenetic factors in the development of AMD, use of antioxidant supplements has been regarded as an effective management strategy of AMD. Antioxidant supplement consumption, including polyunsaturated fatty acids (PUFAs) intake, has been postulated to be a protective factor of AMD . Evidenclate AMD . As we klate AMD , and fislate AMD ,10. Baselate AMD ,12, caselate AMD , and cohlate AMD , fish inlate AMD . In addilate AMD .Meta-analyses, which are a useful statistical tool, could pool the relevant, but independent, studies together and, thus, come to a more powerful conclusion. Meta-analysis was also used in the detection of potential risk factors for AMD. For instance, based on a combination of five prospective cohort studies, Chong et al. found that heavy alcohol consumption was associated with an increased risk of early AMD . For theThis current study was based on eligible observational studies, and the meta-analysis was conducted according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) and Meta-analysis Of Observational Studies in Epidemiology (MOOSE) guidelines ,19. A coThe studies that met the following criteria were considered for inclusion in this meta-analysis: (1) the effect of fish consumption on the risk of AMD was reported; (2) results from prospective cohort studies; (3) the values of relative risk (RR) or odds ratio (OR) with 95% confidence intervals (CI) were provided.Data were independently extracted by two authors (Wei Zhu and Yan Wu) and any disagreements were resolved through discussion with a third author (Yi-Fang Meng). The following data were extracted from each included publication: First author, year of publication, name of cohort, country, age and gender of participants, amount of cases and cohort participants, subtypes or processing methods of fish, adjusting status of the confounding factors, and OR/RR values with 95% CI.Methodological quality assessment of each included study was assessed by two authors (Wei Zhu and Yan Wu). The assessment scores were checked, and any discord was discussed and a unanimous result was obtained. Considering that all the included studies were cohort studies, the Newcastle-Ottawa Scale (NOS), which was designed for the assessments of observational studies, was used in the assessment of the methodological quality of the included studies . The max2 and I2 methods were used in the assessment of heterogeneity in this study. The inter-study heterogeneity was considered statistically significant if p < 0.1 or I2 > 50%. A random-effects model was used in the estimation of the pooled effects when the inter-study heterogeneity was statistically significant. The effects of fish consumption on AMD risk were delineated with RR and a 95% CI. To conduct sensitivity analyses, we dropped included studies, one-by-one, and observed the modification to the conclusion.Both OR and RR were extracted from the included studies and used in the final quantitative synthesis. Considering the relative low incidence of AMD, OR values could be used to approximate RR. The adjusted OR/RR values were adopted in the meta-analysis if possible. Both χp value for nonlinearity was detected by testing the null hypothesis that the coefficient of the second spline is equal to 0 [A two-stage, random-effect, dose-response meta-analysis was conducted for the detection of a potential linear relationship between fish consumption and risk of AMD incidence. Restricted cubic splines with four knots, at percentiles of 5%, 35%, 65%, and 95% of the distribution, were used to examine the potential linear dose-response relationship. A ual to 0 ,22.p value < 0.05 was regarded as statistically significant. All analyses were conducted with STATA statistical software .Publication bias was assessed using two different methods: Visually evaluating a funnel plot and the quantitative Egger test. A A total of 1420 records  were identified through searching the electronic databases. Additionally, 18 more studies were identified through reviewing the reference lists of relevant reviews. A total of 545 unrelated papers were excluded, and 165 publications were reviewed for potential inclusion. After excluding 134 reviews, reviews, case reports, and other articles that reported overlapped data, a total of 31 full texts were assessed for eligibility. Subsequently, a total of 23 studies  were excluded from inclusion, and a final total of eight cohort studies were included for quantitative synthesis ,28,29,30A total of 4202 cases with 128,988 individuals from eight cohort studies were identified in this meta-analysis. The detailed characteristics of each included study are presented in The methodological quality of each included study was detected using the NOS scale. NOS was obtained in order to assess the selection, comparability, and outcome of the cohort studies. The scores of each evaluation of all studies are shown in The pooled estimation on fish consumption and risk of AMD showed that fish consumption can reduce the incidence of AMD. In this meta-analysis, the meta-analyzed RR was 0.76  when any AMD was considered A. Subgron = 7; RR, 0.75; 95% CI, 0.63–0.89), but not the hospital-based group . When the geographical distributions of the included studies were considered, the studies that were conducted in the USA  and Iceland  showed statistically significant results; however, no significant results were detected in Australia  or in the Netherlands . When stratified by the follow-up duration, fish consumption was a protective factor of AMD in both, over 10 years  and less than 10 years  follow-up durations. All the results of the subgroup analyses are presented in Stratified analysis was conducted by data source, study site, and follow-up duration. In the two data source subgroups, a significant association was detected in the population-based group , especially tuna fish  was associated reduced AMD risk. However, no significant association between fish intake and AMD incidence was detected in neither other dark meat fish group  nor non-dark meat fish group . Subgroup analysis using the processing methods showed that no protective effects were detected in backed, fried or smoked fish group. The results of the detailed stratified analyses are presented in I2, 49.6%; p = 0.053). When subgroup analysis by AMD subtypes was conducted, no significant heterogeneity was detected in both groups. When the heterogeneity was significant in the subgroup analysis, a random-effects model was obtained to assess the pooled effect.Heterogeneity was not significant when all eight studies were pooled in the meta-analysis . A one-time-per-week consumption of fish conferred a RR decrease of 0.11 .No significant publication bias in the current meta-analysis was detected using either Begg’s graph or Egger’s test. The funnel plot was symmetrical on visual inspection . In the A total of 4202 cases with 128,988 individuals from eight cohort studies were identified in this meta-analysis. All the included studies demonstrated a relatively high methodological quality. The findings of this meta-analysis indicated that fish consumption was associated with a reduced risk of AMD. Meanwhile, subgroup analysis by AMD stages showed that fish consumption could reduce the risk of both early and late AMD. When stratified by the follow-up duration, fish consumption was a protective factor of AMD, in both over and less than a 10-year follow-up duration. We also detected an association between different types of fish and risk of AMD. Advanced subgroup analysis showed that dark meat fish and tuna fish showed a protective effect on AMD. In addition, obvious evidence of a statistically significant dose-response relationship between fish intake and AMD risk was detected.n-3 PUFA consumption [n-3 PUFA intake and AMD risk showed that higher n-3 PUFA intake could reduce AMD risk [n-3 PUFA, a nutritional biomarker of n-3 PUFA status, was reported to be associated with the incidence of AMD. In a population-based study on nutrition and age-related eye diseases, performed in 963 residents of Bordeaux (France) aged ≥73 years [n-3 PUFAs are associated with a decreased risk for late AMD.It was reported that inflammation and oxidative stress were key pathologic processes in the development of AMD . Those tsumption ,35. A meAMD risk . Plasma n-3 PUFA in blood [Fish, especially tuna fish, is the main source of PUFAs, and higher fish consumption can increase the concentrations of in blood . In thisin blood . Anotherin blood . ConsideIn this meta-analysis, we found that it was tuna fish but not other types of fish that could reduce the risk of AMD. This was also found in the Nurses’ Health Study (NHS) and the Health Professionals Follow-up Study, a significant inverse association was found only in tuna group. The pooled RR of participants who ate canned tuna more than once 4 times per week was significantly lower  . Tuna fin-3 PUFA supplementation in people with AMD does not increase the progression or development of AMD [n-3 PUFA level had an increased risk for prostate cancer [n-3 PUFA supplementation might produce certain harmful effects on chronic inflammation, and a possible explanation for this relates to the fact that polyunsaturated fatty acids act as a substrate for reactive oxygen damage. Dark meat fish, which was the richest source of (docosahexaenoic acid) DHA and (eicosapentaenoic acid) EPA, was associated with reduced AMD risk in this meta-analysis. Thus, additional well-designed studies are required for the detection of the protective effects of anti-oxidants in early AMD.Several previous trials were conducted in order to explore the effects of PUFA supplementation on the prevention of AMD. The Nutritional AMD Treatment 2 Study was conducted to evaluate the efficacy of DHA-enriched oral supplementation in preventing exudative AMD . The stut of AMD . As repoe cancer . It was There are several strengths in this current meta-analysis: (1) A relative comprehensive literature search strategy was used in the search for related publications. We searched databases, including the key words “life style” OR “dietary factor” to detect all available studies; (2) Only prospective cohort studies were included in this meta-analysis, and all included studies demonstrate a relatively high quality. Thus, no significant selection bias influences the conclusion of this study. Robust conclusions were proven through detailed sensitivity analysis and, thus, it suggests that the conclusions of this study are quite credible; (3) A dose-response analysis was conducted and we detected a dose-response relationship between fish intake and AMD risk. The advanced analyses using available data could provide a better understanding of the effect of fish consumption on the risk of AMD.As with any meta-analysis of observational studies, our study has several limitations. Firstly, the amount of included studies was small. Even through a comprehensive literature search was conducted, only eight studies were included in this meta-analysis. This limited the dependability of subgroup analysis, as only a few studies were included. Secondly, most studies did not provide data stratified by some important confounding factors, such as tobacco smoking and family history. Although all the RR values of the included studies were adjusted by key factors, the influence of these factors should not be ignored. These points all indicate the requirement of additional well-designed studies in the future.In conclusion, the results from this meta-analysis of prospective cohort studies demonstrated that fish consumption, especially tuna fish, could reduce AMD incidence. There was a significant dose-response relationship between fish consumption and risk of AMD. However, additional longitudinal studies with more detailed data, such as fish subtypes or processing methods, are still required and would provide a better understanding on this issue.


Sensitivity analysis and publication bias were also performed. For the association between retinol intake and total fracture risk, we performed subgroup analysis by sex, region, case ascertainment, education level, age at menopause and vitamin D intake. R software was used to complete all statistical analyses. A total of 319,077 participants over the age of 20 years were included. Higher dietary intake of retinol and total vitamin A may slightly decrease total fracture risk  and 0.94 , respectively), and increase hip fracture risk  and 1.29 , respectively). Lower blood level of retinol may slightly increase total fracture risk ) and hip fracture risk ). In addition, higher β-carotene intake was weakly associated with the increased risk of total fracture ). Our data suggest that vitamin A intake and level may differentially influence the risks of total and hip fractures. Clinical trials are warranted to confirm these results and assess the clinical applicability.This meta-analysis evaluated the influence of dietary intake and blood level of vitamin A  on total and hip fracture risk. Cohort studies published before July 2017 were selected through English-language literature searches in several databases. Relative risk (RR) with corresponding 95% confidence interval (CI) was used to evaluate the risk. Heterogeneity was checked by Chi-square and I T-score of <–2.5 SD) [Osteoporosis, defined as a bone mineral density 2.5 standard deviations (SDs) or more below the average value of young healthy women (a –2.5 SD) , is one –2.5 SD) . Among t–2.5 SD) ,5. In 20–2.5 SD) . The ann–2.5 SD) . In addi–2.5 SD) .Vitamin A (VA) is a generic term for compounds with the biological activity of retinol . PreformOur meta-analysis was performed in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement .A wide range of original research and observational studies published before January 2017, examining the association between VA and osteoporosis and its related fracture, were searched within the following databases: Medline, Embase, Science Online, Science Direct, JAMA and Nature. Combinations of at least two of the following key words: “osteoporosis”, “fracture”, “bone density”, “bone health”, “bone rarefaction” and “VA”, “retinol”, “β-carotene”, were used as search terms in the title/abstract of articles. Articles that met the inclusion criteria were retrieved for full-text review, unless the article was not available after all attempts to retrieve. In addition, we manually searched relevant publications by screening the bibliography of selected articles (published during 2014–2016), relevant reviews, and meta-analyses. These articles were not included unless original data were found. To reduce the potential selection bias, each article was independently evaluated by two of the investigators , and a mutual decision was made jointly in regard to whether or not it met the basic inclusion criteria. The disagreements were solved by the opinion of the third investigator (Xinge Zhang), and consensus was reached after discussion.The inclusion criteria were listed as follows: (1) subjects were limited to adults; (2) original and cohort studies published in English-language refereed journals; (3) the exposure of interest was dietary or circulating VA ; (4) the outcome was risk of total or hip fracture; (5) estimation of odds ratio (OR), RR or hazard ratio (HR) with 95% CI was provided.Data extraction was conducted using standard data extraction form by one investigator (Guiyu Qin) and checked by another one (Xinge Zhang). Extracted information included: title, last name of the first author, country, follow-up period, number and age of subjects, measure or range of exposure, research endpoint, case ascertainment, HR/OR/RR with 95% CI, variables adjusted for analysis.We used the Newcastle-Ottawa Scale (NOS) to assess the quality of included studies . The spehttp://getdata-graph-digitizer.com/) was applied to digitize and extract the data. Standard errors were calculated from 95% CIs with a formula [The measures of effects were RR, OR and HR with their corresponding 95% CIs. We treated the HR as RR and we converted OR into RR according to the method by Zhang et al. . Some st formula . Some stFor studies researching VA  intake and risk of total or hip fracture, we extracted, pooled and synthesized values and corresponding 95% CI of the highest versus the lowest category of retinol intake. Subgroup analyses stratified by sex, region, case ascertainment, education level, age at menopause and vitamin D intake were conducted for meta-analysis of retinol intake and total fracture. The random effects model  was usedFor blood level of VA  and risk of total or hip fracture, we extracted estimate values with their corresponding 95% CIs of both the highest and lowest values versus middle category. There were too few studies concerning β-carotene, total VA and fracture to carry out a meta-analysis, we therefore only analyzed the relationship between blood retinol and total or hip fracture.2 test [p value < 0.1 and I2 > 50% indicated high heterogeneity. For sensitivity analysis, we omitted one study each time and then calculated the pooled RR and 95% CI to identify the studies that included potential confounding factors. Funnel plots and Egger’s linear regression method were used to evaluate potential publication bias [The statistical heterogeneity was assessed by Chi-square and I2 test . p valueion bias . All staAfter the title and abstract screening followed by the full-text screening, a total of 13 articles (11 prospective cohort studies and 2 nested case-control studies) published between 2002 and 2015 were included based on our inclusion criteria. Characteristics of eight studies investigating the relationship between VA  intake and fracture risk  ,26,35,36Characteristics of five studies regarding the relationship between blood retinol level and risk of fracture  are summarized in 2 = 35.18%, p = 0.20) or hip fracture .Three studies were included for the analyses of the relationship between total VA intake and total fracture ,19,21 wh2 = 64.64%, p = 0.04), but not in that between retinol and total fracture .Five and four studies respectively reported the relationship between retinol intake and total fracture and the relationship between retinol intake and hip fracture were included ,21,26,352 = 0.00%, p = 0.52), but not that between β-carotene intake and hip fracture   .2 = 0.00%, p = 0.64) or for hip fracture   ,22,37,38 = 0.62) .Subgroup analyses were performed to determine the relationship between retinol intake and total fracture risk, and the results were shown in p = 0.005)  .Our results suggested that higher intake of total VA or retinol slightly increased the risk of hip fracture. A number of cohort studies indicated a similar trend  ,21, highIn addition, the interactions between retinol and vitamin D or calcium may play a role. Excessive retinol intake can alter the metabolism of calcium-regulating hormones  and elimOur findings are not consistent with the results reported in a previous meta-analysis conducted by Wu et al. . The autSubgroup analysis for retinol intake and total fracture suggested that sex, age at menopause, case ascertainment and region might be potential factors, which were not reported in a previous meta-analysis by Wu et al. . It has Here. A positive association between β-carotene intake and risk of total fracture was found. Interestingly, Wu et al.  reportedPrevious study reported a U-shaped relationship between serum retinol and hip fracture risk . ConsistThis meta-analysis has several potential limitations. First, we failed to obtain the exposure categories and RR with 95% CI of our interests in the study conducted by Sahni et al. , which wOur meta-analysis suggested that higher total VA or retinol intake might slightly decrease the risk of total fracture and increase the risk of hip fracture. In addition, we found that higher β-carotene intake increases the risk of total fracture slightly but not hip fracture. Moreover, we observed weak but positive influence of lower blood retinol level on both total and hip fracture risk.


NRG1) as a critical component in the pathophysiology of schizophrenia. Although a number of risk haplotypes along with several genetic polymorphisms in the 5′ and 3′ regions of NRG1 have been linked with schizophrenia, results have been mixed. To reconcile these conflicting findings, we conducted a meta-analysis examining 22 polymorphisms and two haplotypes in NRG1 among 16 720 cases, 20 449 controls and 2157 family trios. We found significant associations for three polymorphisms  at the 5′-end and two (rs2954041 and rs10503929) near the 3′-end of NRG1. Population stratification effects were found for the rs35753505 and 478B14-848(4) polymorphisms. There was evidence of heterogeneity for all significant markers and the findings were robust to publication bias. No significant haplotype associations were found. Our results suggest genetic variation at the 5′ and 3′ ends of NRG1 are associated with schizophrenia and provide renewed justification for further investigation of NRG1’s role in the pathophysiology of schizophrenia.Genetic, post-mortem and neuroimaging studies repeatedly implicate neuregulin-1 ( The gene encoding NRG1 is large, spanning ~1.2 Mb and contains >23 000 single-nucleotide polymorphisms (SNPs) among which ~40 have been associated with schizophrenia.5 Genome-wide association studies have generally, however, only provided modest support with the most recent study implicating rs986110 (P=1.5 × 10−4) with the disorder.6 This may in part be due to genome-wide association study to date focussing exclusively on SNP variation and consequently underestimating the importance of genes, such as NRG1, for which haplotype and microsatellite variation has been demonstrated. Thus, arguably a more thorough evaluation of NRG1’s association with schizophrenia requires examination of variation beyond SNPs.Neuregulin-1 (NRG1) is a pleiotropic growth factor involved in circuitry generation, axon ensheathment, neuronal migration, synaptic plasticity, myelination and neurotransmission.NRG1 primarily sit within untranslated or intronic regions at the 5′ and 3′ ends of the gene. Yet, research to date has focused on the 5′-region of NRG1. This 5′-bias has been driven by the landmark study in 2002 conducted by Stefansson et al.,7 who identified a seven-marker schizophrenia-associated haplotype in the Icelandic population (HapICE) consisting of five SNPs and two microsatellites (478B14-848 and 420M9-1395) in the 5′-region of NRG1. As this milestone study, additional 5′-schizophrenia-associated haplotypes in the Irish (HapIRE)8 and Chinese (HapChina1-3)9 populations have been identified. However, the most recent meta-analysis conducted in 2008  by searching three electronic databases: PubMed, PsychInfo and Medline (Ovid), using the search terms ‘neuregulin 1’, ‘neuregulin-1’, ‘neuregulin1’, ‘schizophrenia’ and ‘association’, and the abbreviation of the gene ‘NRG1’ and ‘NRG 1’ with no language restrictions. Bibliographies of all research articles were hand searched for additional references not indexed by MEDLINE. In cases where genotype data were not available in the published research articles or www.szgene.org) as a resource for collecting genotype data. All publications published from January 2002 through February 2016 were assessed for inclusion.The 2015 PRISMA-P  checklistNRG1; (b) published in peer-reviewed journal containing original data; (c) included clinically diagnosed schizophrenia patients using an accepted classification system ; and (d) provided sufficient genotype or allelic data for calculation of an odds ratio (OR). Based on these criteria, 48 (40 case–control and 8 family-based) studies were included  a case–control or family-based genetic association studies investigating one or more SNPs and/or microsatellites of included .From each case–control and family study, the following data were extracted: (a) author(s) and publication year; (b) number of cases and controls or family sample size; (c) country of origin or ethnicity; (d) diagnostic criteria used; (e) SNP reference sequence number or marker identifier; (f) the publication identification number ; (g) genotype counts and/or allele counts in cases and controls or family samples; and (h) haplotype frequencies in cases and controls (where available). Extracted data for all selected studies can be found in Data from each case–control study were used to create 2 × 2 tables and data from each family study were used to create 1 × 2 tables. Classifications of the subjects were based on diagnostic category and type of allele they carried.R version 3.3.0 . The meta12 and metafor13 packages were used to conduct the meta-analyses. The OR with 95% confidence intervals (CIs) was used as the effect size estimator. The method proposed by Kazeem and Farrall14 was used to calculate the effect size for transmission disequilibrium test studies, where the ORs were estimated from the number of transmissions versus non-transmissions of the designated high-risk allele to schizophrenia cases from heterozygous parents. For case–control studies, ORs were estimated by contrasting the ratio of counts of the high-risk versus low-risk alleles in schizophrenia cases versus non-clinical controls. For those polymorphisms in which the previous literature provided an indication of the risk-inducing allele, one-tailed P-values were reported. In the absence of prior data, two-tailed P-values were reported and were indicated accordingly in the text. All statistical tests (except for the Q-statistic) were considered statistically significant at P<0.05.Data were analysed using 15 which is based on a normal distribution. The standard error estimates were adjusted using the Hartung–Knapp–Sidik–Jonkman17 correction, which then calculates the corresponding 95% CI based on the t-distribution. The Hartung–Knapp–Sidik–Jonkman method generally outperforms the DerSimionian–Laird approach on type-I error rates when there is heterogeneity and the number of studies in the meta-analysis is small.19Because of the differences in study design and sample characteristics, considerable heterogeneity was expected between the studies. Therefore, the pooled OR was calculated using the random-effects models with the DerSimionian–Laird estimator,20 Studies with observed effects that are well separated from the rest of the data are considered outliers. Such studies were identified using studentised deleted residuals, with absolute values >1.96 indicative of outliers. An influential study leads to considerable changes to the fitted model and a range of case deletion diagnostics adapted from linear regression can be used to identify these studies, including the DFFITS, DFBETAS and COVRATIO statistics (see Viechtbauer and Cheung20 for more information). Potential outliers and influential studies were omitted and the analyses were then re-run to determine their influence on the pooled effect size.Outliers and influential studies were identified according to the recommendations of Viechtbauer and Cheung.Q-statistic (with P<0.10 indicating significant heterogeneity) and its magnitude was quantified using the I2 statistic, which is an index that describes the proportion of total variation in study effect size estimates that is due to heterogeneity and is independent of the number of studies included in the meta-analysis and the metric of effect sizes.21 As the Q-statistic has low power when the number of studies is small,22 95% prediction intervals were calculated to quantify the extent of heterogeneity in the distribution of effect sizes.23 The prediction interval is an estimation of the range within which 95% of the true effect sizes are expected to fall.Heterogeneity in effect sizes across studies was tested using the 24 which estimates the number of studies missing from the funnel plot and imputes these missing studies to make the funnel plot symmetrical, and then calculates an estimate of the effect size adjusted for publication bias.25 Following the recommendations of Sterne et al.,26 a test for funnel plot asymmetry was only conducted if the number of studies was 10 or greater. The regression test proposed by Harbord et al.27 was used to quantify the bias captured by the funnel plot and tested whether it was statistically significant. In addition, cumulative meta-analyses sorted by the sampling variance of the respective studies were conducted to examine the relationship between imprecise samples and effect sizes.28 This visualises the effect that small imprecise study samples have on the estimations of the pooled effect size.Publication bias was assessed using funnel plots and the trim-and-fill procedure,29 was used for the haplotype meta-analyses to avoid the inflation of the type-I error rate that is observed in the traditional approach of comparing a haplotype against the remaining ones.29The generalised linear mixed model method  detailed in BagosModerator analyses for study design, diagnostic criteria and ancestry were conducted using mixed-effects meta-analyses. For this method, studies within potential moderator groups were pooled with the random-effects model, whereas tests for significant differences between the groups were conducted with the fixed-effects model. The Hartung–Knapp–Sidik–Jonkman adjustment was used if there were at least three studies in each group, otherwise the unadjusted DerSimionian–Laird method was used.A total of 22 single markers and two haplotypes that appeared in three or more studies were examined . SignifiI2=18.5–54.3%). The funnel plots are presented in k=5, OR=1.14, 95% CI=1.10–1.18, P⩽0.001).Across the three significant single markers, heterogeneity was low to moderate  and conversely a ‘protective’ association among European studies  but not European  studies ,33 with the exception of the 3′ SNP rs2954041. To our knowledge, this is the first meta-analysis to identify an association between schizophrenia and rs2954041.Three of the seven HapICE markers  at the 5′-end as well as two SNPs (rs2954041 and rs10503929) near the 3′-end of NRG1, ~18 kb from the type III (SMDF) promoter, the most brain abundant isoform of NRG1.34 To our knowledge, rs2954041 has not been assessed as expression quantitative trait loci for type III expression. However, given its proximal location to the type III promoter and preclinical evidence suggesting disruption of type III results in phenotypes commonly associated with schizophrenia ,35 rs2954041 could have a functional role in the pathophysiology of schizophrenia. In addition, others have shown this SNP interacts with rs7424835 in ERBB4, the cognate receptor for NRG1  and are based exclusively on studies within populations of European descent. This is notable because the rs10503929 ‘risk’ allele (T) is the major allele and is carried by all East Asians, 99% of Africans and 94% of South Asians relative to 81% of Europeans (http://browser.1000genomes.org/index.html). Thus, future studies in Asian and/or African populations may not be relevant or will require extremely large sample sizes.The rs2954041 SNP is located in the fifth intron of RG1 ref.  further NRG1 that associate rs62510682, rs35753505 and 478B14-848 with schizophrenia have previously been identified in other meta-analyses. The rs35753505 is the most studied and the first NRG1 marker to receive meta-analytic support for an association with schizophrenia.30 However, in three subsequent meta-analyses, this association was not detected.32 In the current meta-analysis, we have revived this association but only among Asians, which is contrary to the original meta-analytic association for rs35753505 that was found only among Caucasians.30 This finding is perhaps not surprising given evidence of population stratification at the NRG1 locus.10 In fact, we also found that the 4 allele of the 478B14-848 microsatellite is a marker of ‘risk’ among Asians but ‘protection’ among Caucasians. This aligns with knowledge that the 0 allele in Asian populations is low39 compared with the 4 allele, which is quite prevalent and forms in part the HapCHINA schizophrenia risk haplotype.40 However, no other markers we investigated were moderated by ancestry, including the three omnibus markers  and rs2954041) associated with schizophrenia, albeit the number of non-Caucasian studies available for many of the markers hinders firm conclusions.Our findings from the 5′-end of NRG1 marker but previous meta-analyses have been mixed. Li et al. showed in a meta-analysis of eight studies that carriers of the G allele had greater odds of a schizophrenia diagnosis, particularly among individuals of European descent; but in a subsequent meta-analysis of 14 studies by Gong et al., this association was not upheld. Our meta-analysis of rs62510682 included 25 studies, a near doubling of the most recent meta-analysis, and reproduced the finding reported by Li et al. that suggests the G allele of rs62510682 is associated with schizophrenia. Our moderator analysis showed that this association did not differ by ancestry, although stratification analysis did suggest that this association might be stronger among individuals of European descent.The rs62510682 (SNP8NRG241930) is the second most frequently studied 30 although Li et al. combined carriers of the 0 and 4 alleles in their meta-analysis—an approach that has important implications with interpretation given our finding that the 4 allele can confer a ‘risk’ or ‘protective’ effect depending on ancestry. Nevertheless, our meta-analysis results uphold the meta-analytic association between the 0 allele and schizophrenia reported by Gong et al. and support further study of this potentially important microsatellite.Although studied less frequently than other HapICE markers, the 0 ‘risk’ allele of the microsatellite 478B14-848 has been linked to schizophrenia in two previous meta-analyses,10 they pooled the results for non-identical five- and seven-marker haplotypes. Thus, their results do not reflect the overall association of the HapICE haplotype block in schizophrenia. Furthermore, most of the included studies were conducted in populations of European ancestry, which is not surprising given the frequency of the alleles that constitutes the HapICE risk haplotype is relatively low in Asian populations. In fact, most Asian studies do not look at the full HapICE haplotype but rather select SNPs and microsatellites forming the HapCHINA haplotype.Our results, however, do not support an association between either the five- or seven-marker HapICE haplotypes and schizophrenia. To our knowledge, this is the first meta-analysis to examine the five- and seven-marker HapICE haplotypes. Although previous meta-analysis have showed positive associations for both five- and seven-marker haplotypes in schizophrenia,NRG1. Although support for an association between the five- or seven-marker HapICE haplotypes and schizophrenia was not found, three of the markers within these haplotypes had robust associations. Our results highlight the importance of genetic variation at both the 5′ and 3′ ends of NRG1 and provide justification for further investigation of NRG1’s role in the pathophysiology of schizophrenia.In conclusion, we have replicated and identified novel strong positive associations among polymorphisms situated at the 5′ and 3′ ends of


Urothelial Carcinoma Associated 1 (UCA1) was an originally identified lncRNA in bladder cancer. Previous studies have reported that UCA1 played a significant role in various types of cancer. This study aimed to clarify the prognostic value of UCA1 in digestive system cancers.The meta-analysis of 15 studies were included, comprising 1441 patients with digestive system cancers. The pooled results of 14 studies indicated that high expression of UCA1 was significantly associated with poorer OS in patients with digestive system cancers . In addition, UCA1 could be as an independent prognostic factor for predicting OS of patients . The pooled results of 3 studies indicated a significant association between UCA1 and DFS in patients with digestive system cancers . Statistical significance was also observed in subgroup meta-analysis. Furthermore, the clinicopathological values of UCA1 were discussed in esophageal cancer, colorectal cancer and pancreatic cancer.A comprehensive retrieval was performed to search studies evaluating the prognostic value of UCA1 in digestive system cancers. Many databases were involved, including PubMed, Web of Science, Embase and Chinese National Knowledge Infrastructure and Wanfang database. Quantitative meta-analysis was performed with standard statistical methods and the prognostic significance of UCA1 in digestive system cancers was qualified.Elevated level of UCA1 indicated the poor clinical outcome for patients with digestive system cancers. It may serve as a new biomarker related to prognosis in digestive system cancers. Cancer is now becoming the leading cause of death in both developed and developing countries . DigestiLncRNAs were defined as non-protein coding RNAs with the length of more than 200 nucleotides. LncRNAs account for more than 80% of the entire genome transcripts . Over thUrothelial carcinoma associated 1 (UCA1) was a novel lncRNA gene with three exons and two introns, which is located in the chromosome 19p13.12. The lncRNA UCA1 was firstly discovered to be up-expressed in the carcinogenesis of bladder cancer in 2006 . In addiAccording to the criteria for selection mentioned above, after carefully screening the abstract and full-text of these references, finally, 14 publications (including 15 studies) –29 were A total of 1441 patients with survival data were included in this meta-analysis, the mean sample size was 96.07, with a maximum number of 240 and a minimum sample size of 20. Among the fifteen studies, six studies were about colorectal cancer , 24, 27,P = 0.995, I2 = 0.0 %). Overall, the pooled results confirmed that there was a significant association between high UCA1expression and poor OS in digestive system cancer   Figure .Although there was no significant between-studies heterogeneity, the subgroup meta-analysis was conducted on the tumor type, country, histology type, analysis type, sample size and cut-off value Table . From thThe sensitivity analysis was performed to assess the stability of the results by removing each study in turn. The result indicated that meta-analysis results didn't not change significantly Figure , which s2= 0.0%), the fixed-effects model was utilized to combine the hazard ratios (HRs) with corresponding 95 % CI. The pooled meta-analysis confirmed that high UCA1 expression was a significant independent predictor of poor OS in digestive system malignancies   Figure . The patMoreover, the pooled HR values > 1 were consistently calculated in subgroup meta-analysis stratified by the tumor type, histology type, sample size and cut-off value, which still had statistical significance Table . The subThe leave-one-out sensitivity analysis performed showed that no individual study changed the pooled HRs significantly Figure . However2 = 0.0 %; P = 0.994), the fixed-effects model was used to analyze the pooled hazard ratios (HRs) with corresponding 95 % confidence interval (CI). The results showed that there was a significant association between high expression level of UCA1 and poor DFS in digestive system malignancies  , age  and TNM stage , the fixed-effects model was used for all , location , differentiation grade , lymphatic invasion, venous invasion , lymph node metastasis , distant metastasis , and TNM stage  in colorectal cancer revealed no significant heterogeneity across studies, and thus the fixed-effects model was applied. However, there were significant heterogeneity in studies regarding depth of primary tumor invasion  and tumor size  , lymph node metastasis  and distant metastasis and TNM stage . Whereas no significant association was found with gender , tumor size , location , the depth of tumor, lymphatic invasion  and venous invasion . Publication bias and sensitivity analysis were not performed due to the small number of studies and little heterogeneity.In pancreatic cancer, there were only two studies assessing the correlation between UCA1 expression and clinicopathological data , 28, the2%= 67%), there was no significant heterogeneity between studies in gender , location , nervous invasion  and depth of invasion . Thus, the fixed-effects model was used for all except for the histological grade , location , histological grade  and nervous invasion  . The patients with high UCA1 expression had a poorer OS compared to patients with low UCA1 expression. Despite significant heterogeneity didn't exist in these studies for OS, subgroup analyses were performed on the tumor type, country, histology type, analysis type, sample size and cut-off value  . And no obvious heterogeneity was observed in subgroup analysis of studies. The prognostic implication of UCA1 was also showed in the stratified analysis based on cancer type, histology type, sample size and cut-off value Table . None ofet al. [et al. [et al. [For the relationship between UCA1 and clinicopathological features in digestive system cancers, multiple studies have been performed on different kinds of the cancers. But clinicopathological value of UCA1 in a specific kind of digestive system cancer was distinctive and even contradictory. For example, in human colorectal cancer, Han et al.  suggeste [et al. indicated [et al.  reportedThe molecular biology characters were different and mechanism of tumorigenesis was complex in the digestive tract tumor. Thus, we focused on the clinicopathological value of UCA1 in a specific type of digestive tract cancers. We studied the association between UCA1 expression and clinicopathological characteristics in esophageal cancer, colorectal cancer and pancreatic cancer. The clinicopathological value of UCA1 in gastric cancer and hepatocellular carcinoma was not obtained for the limited and unavailable data of clinical pathological features. In esophageal cancer, we found UCA1 was significantly associated with advanced TNM stage, but no significant association was found in gender and age. In colorectal cancer, our research demonstrated that UCA1 expression was significantly related to differentiation grade, lymph node metastasis, distant metastasis and TNM stage, but no significant association was observed in gender, tumor size, location, the depth of tumor, lymphatic and venous invasion. Finally, the clinicopathological role of UCA1 was evaluated in pancreatic cancer, UCA1 expression was significantly associated with primary tumor invasion, however no significant association was found in gender, location, histological grade and nervous invasion.et al. [et al. [et al. [The expression patterns and biological roles of UCA1 in digestive system cancers was explored. It would further support UCA1 as a promising biomarker for the prognosis of digestive system tumors. The potential roles of UCA1 expression regulation in various digestive carcinomas have been reviewed. For gastric cancer, UCA1 was observed to be highly expressed, while the silence of UCA1 could decrease proliferation of tumor cells. Expression of UCA1 was negatively correlated with the miR-27b and the UCA1-miR-27b axis was involved in regulation of chemo-sensitivity of gastric cancer cells . In esopet al.  identifiet al. . For the [et al.  detected [et al.  also proIt should be emphasized that the present meta-analysis had several limitations. First, the criterion of high expression for UCA1 in tissue samples was not the same in different studies, and it was hard to get a consensus cut-off value to define the UCA1 overexpression in various cancers. Second, the therapeutic strategy was diverse in gastrointestinal tumors, which made a significant influence on postoperative survival of patients, leading to heterogeneity. Third, we only studied publications written in English and Chinese, and only 15 studies with 1441 patients were included in present meta-analysis, so the total number of studies and patients included was relatively small. Fourth, only 3 studies including 429 patients were included and reported the relationship between UCA1 and DFS. Furthermore, all the studies we included were conducted in Asian population from Chinese and Korea. Finally, the potential publication bias was also observed in this meta-analysis and positive results would be more easily to be published than that of in negative results.In conclusion, the meta-analysis results of this study could help to improve our understanding on the prognosis significance of UCA1 in different types of digestive system carcinomas. UCA1 may serve as a novel biomarker for predicting the prognosis and assessing clinicopathologic features in digestive system carcinomas. Ultimately, further scientific research with larger-size, multi-center and higher-quality studies is required to verify the clinical utility of UCA1 in digestive system malignancies.To get access to potentially eligible studies, comprehensive literature retrieval was performed against several databases: PubMed, Web of Science, Embase and Chinese National Knowledge Infrastructure (CNKI), Wanfang database. The literature search was conducted up to Oct. 15, 2016. The publications were identified with the combination of the following search terms: “Urothelial cancer-associated 1” or “UCA1” or “lncRNA UCA1” or “long noncoding RNA UCA1”, “cancer” or “carcinoma” or “tumor” or “neoplasm,” “prognosis” or “survival” or “clinical outcome”. The reference lists of relevant articles were also searched manually. The published language was limited to English and Chinese.Inclusion and exclusion criteria were set to screen the publications. Inclusion criteria were as follows: (1) The studies explored the association between UCA1 and cancer prognosis; (2) Related clinicopathologic parameters were described; (3) The UCA1 expression were determined in the tissues of digestive system cancers; (4) Patients were divided into high and low expression groups according to the expression level of lncRNA UCA1; (5) Sufficient information and data was provided for calculating a hazard ratio (HR) with its 95% confidence interval (CI). Exclusion criteria were as follows: (1) The studies not relevant to digestive system cancers, UCA1, or cancer prognosis; (2) studies without usable data, such the data was obtained from animal models; (3) duplicate publications; and (4) reviews, letters and case reports.Two authors (DQ and GH) extracted data independently from all the including studies by standardized data compilation forms, and any disagreements were resolved by discussion. For all eligible studies, the following information was collected: the first author, year of publication, country, cancer type, sample size, tumor stage, follow-up period, outcome measures, cut-off value, determination method, HR and corresponding 95 % CI. The clinicopathological data were also extracted from the eligible studies.http://digitizer.sourceforge.net/).If the HRs with corresponding 95 % CIs was provided in a including study, the available data was directly extracted. If a study provided only Kaplan-Meier curves, the survival data was extracted from Kaplan-Meier survival curves by Engauge Digitizer V4.1  was applied to assess the quality of all including studies. The scores of NOS criteria were ranged from 0 (lowest) to 9 (highest). If the final scores of a study was higher, the methodological quality was better. A study with an NOS score equal or more than 6 was considered to be of high quality. In this meta-analysis, the quality of all studies included in this meta-analysis was varied from 6 to 9, with a mean value of 7.5.Stata Corporation, College Station, TX, USA). ORs for clinicopathological parameters were calculated by RevMan5.3 software .Statistical analyses of HRs for OS and DFS were calculated by Stata SE12.0 ; otherwise, the random effects model was used to provide wider CI for studies with significant heterogeneity.The between-studies heterogeneity was determined by Chi square-based Q test and ISubgroup analysis was performed to further investigate the prognostic value of lncRNA UCA1 in digestive system malignancies. Potential publication bias was determined with funnel plots analysis, Begg's test and Egger's test. The sensitivity analysis was also performed to assess the stability of the results. A p value less than 0.05 was statistically significant.