Cholera has caused 7 global pandemics, including the current one which has been ongoing since 1961. A systematic review of risk factors for symptomatic cholera infection has not been previously published.In accordance with PRISMA  guidelines, we performed a systematic review and meta-analysis of individual and household risk factors for symptomatic cholera infection.I2 = 8%), unimproved water source , open container water storage , consumption of food outside the home , household contact with cholera , water treatment , and handwashing . Other notable associations with symptomatic infection included income/wealth, blood group, gastric acidity, infant breastfeeding status, and human immunodeficiency virus infection.We identified 110 studies eligible for inclusion in qualitative synthesis. Factors associated with symptomatic cholera that were eligible for meta-analysis included education less than secondary level (summary odds ratio [SOR], 2.64; 95% confidence interval [CI], 1.41–4.92; We identified potential risk factors for symptomatic cholera infection including environmental characteristics, socioeconomic factors, and intrinsic patient factors. Ultimately, a combination of interventional approaches targeting various groups with risk-adapted intensities may prove to be the optimal strategy for cholera control. Vibrio cholerae, is endemic to the Indian subcontinent and has caused 7 recorded global pandemics [Cholera, the acute watery diarrheal illness caused by toxigenic andemics . The sevandemics , 3. Sincandemics , 5. As tandemics , it willWe conducted this study in accordance with PRISMA  guidelines, which provide evidence-based recommendations for conducting and reporting systematic reviews and meta-analyses .We searched for peer-reviewed articles assessing individual or household risk factors for symptomatic cholera infection, either with microbiologic confirmation or as defined by the World Health Organization (WHO) case definition of cholera: a patient ≥5 years of age with acute watery diarrhea, with or without vomiting, in an area with a known cholera epidemic . We consVibrio cholerae”) AND (“risk” or “predict” or “risk factor”). We manually reviewed reference lists of related reviews and all included articles.We searched PubMed, Embase, and the Cochrane Library using the following terms:  using random effects models and a generic inverse variance approach to allow for inclusion of odds ratios controlled for measured confounders. We assessed heterogeneity with the Cochran We performed a prespecified sensitivity analysis for risk factors included in meta-analysis by assessing for a subgroup difference between studies with and without microbiologic confirmation of cholera infection. We also evaluated for a subgroup difference between studies that took place in endemic settings compared to epidemic settings.Data were analyzed using Review Manager (RevMan) version 5.3 .The database search was performed on 18 January 2018 and is summarized in The majority of studies were both age- and sex-matched analyses, but a subset assessed risk for cholera based on these characteristics. Fifteen studies assessed differential risk by age , 118, 10I2 = 80% (Q test P < .0001).The role of sex was evaluated in 15 studies , 118, ofSeven studies directly assessed the relationship between income and risk of cholera , 43, 116Seven studies considered socioeconomic status as measured by asset ownership or composite wealth index , 85, 90.Eight studies evaluated household building materials , 110. SeI2 = 8% (Q test P = .35).Fourteen studies explored the role of education , 86, 101There were mixed findings by 8 studies assessing the relationship between number of household members and risk of cholera , 118. OfFour studies found no relationship between cholera and electricity in the household , 86, 118I2 = 77% (Q test P < .00001). Of the 11 studies excluded from meta-analysis, 1 found a significant association but did not report an effect measure [Fifty-six studies evaluated risk of cholera based on source of water, with many identifying a specific culprit , 118. Th measure , 6 used  measure , 105, an measure , 79, 110I2 = 62% (Q test P = .02).Methods of water storage were assessed by 17 studies , 112. OfI2 = 74% (Q test P < .00001). There was no significant difference between the subgroups (P = .65). Of the 12 multivariable analyses excluded from meta-analysis, 8 did not report an effect measure  [Thirty-nine studies considered the relationship between risk of cholera and household water treatment, typically in the form of either chlorination or boiling , 119. Twdid not) , 119. Thdid not) , 49, 86.Distance to a water source was assessed in 5 studies which implemented multivariable models , 32, 33.Two studies found no relationship between availability of water and risk of cholera , 85.Reported exposure to a specific food was commonly assessed. In particular, the role of seafood was evaluated in 23 studies , 119. ImI2 = 64% (Q test P = .0008). Of the 5 controlled studies not eligible for meta-analysis, 3 did not report an effect measure [Twenty-four studies assessed risk of cholera with exposure to food from street vendors or outside the home , 119. Se measure , 92, 119There were mixed results when looking at cholera risk based on hot meal preparation or exposure to leftover food , 118. OfThe possible protective role of breastfeeding in the setting of cholera was first noted in 1979, when bottle-fed children in a matched case-control study had significantly higher risk of cholera . This reV. cholerae in their stool [One study found that retinol deficiency was associated with a higher likelihood of developing symptomatic disease among people growing ir stool , and anoir stool .No studies evaluated risk of cholera based on access to food or food security, although one multivariable analysis found dietary diversity to be associated with a reduced risk of cholera . No studAccess to a flush toilet, latrine, or open defecation was included as a potential risk factor in 27 studies , 107. OfSix studies assessed whether sharing a latrine was a risk factor for cholera , 118. FoI2 = 81% (Q test P < .00001). The 6 studies excluded from meta-analysis used hospital or clinic controls with either noncholera diarrhea or without diarrhea [The risk of cholera with a household contact with cholera was evaluated in 25 studies , 110. Fidiarrhea , 97, 105Two studies looked specifically at sharing a latrine with a person with cholera; 1 found a significantly increased risk and 1 found no significant association , 105.Two studies found that individuals living in close proximity to other cases had the greatest risk of cholera , 71.I2 = 49% (Q test P = .01), with no significant difference between the subgroups (P = .95). Three other studies that included handwashing after defecation in multivariable analyses but did not report effect measures found no significant association with cholera [Hand hygiene was assessed in a number of ways, including handwashing before eating , 116, ha cholera , 37, 79.Of 4 studies evaluating the role of bathing in unsafe water, 2 found increased risk of cholera and 2 found no significant association , 82, 109I2 = 38% (Q test P = .14).Fourteen included studies evaluated risk of cholera after attending a large gathering or funeral , 116. SeV. cholerae [+b–) was associated with symptomatic cholera in 1 study [Multiple studies found higher risk of symptomatic cholera among patients with blood group O , 48, 53,cholerae . Among p 1 study .V. cholera lipopolysaccharide. They went on to conduct a case-control study, finding that these genes were strongly associated with susceptibility to symptomatic cholera.One study further explored the role of genetics in the risk of symptomatic cholera . In a geV. cholerae, from 108 organisms to 104 [Helicobacter pylori immunoglobulin G was associated with risk of cholera [Decreased gastric acid levels have been implicated as a risk factor for cholera infection in several ways. In an experimental trial of cholera inoculation, buffering gastric acid led to a lower required infectious dose of s to 104 . Observas to 104 , 99, 117 cholera . One stu cholera , 69. Ant cholera , 111.I2 = 72% (Q test P = .03).Four multivariable analyses assessed use of acidic additives to food. Three of these reported an effect measure and were candidates for meta-analysis  94, 95,,95, 109.Two case-control studies, 1 using hospital controls and 1 with community controls, found an increased risk of cholera among people with human immunodeficiency virus (HIV) infection , 118.V. cholerae O1 demonstrated natural immunity to clinical infection when rechallenged 3 years later [V. cholerae O1 infection with prior V. cholerae O1 infection, but no significant protection for V. cholerae O139 or cross-protection between serogroups [V. cholerae O1 in household contacts of cholera cases, but not for V. cholerae O139 [A trial of experimental inoculation of volunteers with rs later . A case-rogroups . In 2 strae O139 , 100. Anrae O139 .There were no significant subgroup differences for any of the variables included in meta-analysis when stratifying by whether studies used microbiologic confirmation of cholera cases or a clinical case definition, or whether the studies took place in an endemic setting or an epidemic setting.A summary of bias within studies can be found in This systematic review of 110 studies and >22000 people with cholera identified factors associated with symptomatic infection related to water, sanitation, and hygiene (collectively referred to as WASH), other sources of exposure, socioeconomic status, and intrinsic patient characteristics.Use of an unimproved water source conferred >3-fold increase in the odds of cholera in meta-analysis, highlighting the importance of a safe water source in cholera control. There was significant heterogeneity between the effects of water source on cholera risk among studies included in meta-analysis, probably reflecting differing degrees of contamination or exposure in different water sources. Use of an open container or bucket for water storage was also significantly associated with cholera, potentially through higher risk of water contamination. Water treatment, either by chlorination or boiling, was associated with lower risk of cholera. There was substantial heterogeneity among studies assessing water treatment, possibly reflecting differences in practice or reporting.While access to adequate sanitation is likely to minimize risk of water supply contamination, use of latrines did not convincingly impact an individual’s risk of cholera, with the majority of well-controlled studies assessing individual latrine use finding no association with risk of cholera. Access to and use of hygiene measures, sanitation infrastructure, and sharing may impact cholera risk with latrine use, although none of this was definitively evident from the available information.Handwashing, either before meals or generally, had a similar effect measure for decreased cholera risk as water treatment. This finding is consistent with a recent systematic review of the use of handwashing for prevention of diarrhea more generally . HandwasOther non-water-related sources of exposure were also implicated in the risk of cholera. Food items, and seafood in particular, were associated with cholera in a number of studies. More generally, eating street food was associated with a 5-fold increase in the odds of cholera in meta-analysis, indicating that interventions with street vendors may be an important pathway to interrupt transmission during an outbreak, although current programmatic guidelines on how to address this are nonspecific . SimilarSeveral socioeconomic factors were independently associated with risk of cholera. Both direct income and composite wealth were closely linked to cholera risk in all multivariable analyses assessing these variables. Education  was associated with cholera risk. Taken together, these findings strongly point to the need to explore how poverty reduction, social support, and educational infrastructure could be used as interventions with a goal of cholera control.H. pylori infection or otherwise, decreased the required infectious dose of cholera, thus increasing the risk of symptomatic infection. Finally, HIV infection may increase risk of cholera, although there are only 2 studies assessing this association and both were at high risk for selection bias.Some intrinsic patient factors were also associated with cholera risk. Several genetic features, related to blood group O and genes in the NF-κB pathway, are associated with increased risk of symptomatic cholera, and appear to have contributed to natural selection in parts of Asia. Risk of cholera by age varied substantially between studies and may reflect differences in populations living in epidemic vs endemic areas, where natural immunity may play a role. Breastfeeding infants showed evidence of protection from cholera, probably as a result of passive immunity from breast milk and reduced likelihood of direct exposure. Reduced gastric acidity, from There was evidence of publication bias for some variables included in meta-analysis, including sex, water source, water treatment, water storage, street vendor food exposure, attending a large gathering or funeral, and household cholera contacts. The summary effect measures for these risk factors may thus be biased away from the null because negative associations are less likely to be reported or published. In some cases, studies that would have otherwise been eligible for meta-analysis did not report an effect measure. While generally there were relatively equal numbers of these studies reporting a null or significant association in one direction or the other, this may have introduced some bias into the resulting summary effect measures. We hypothesized that there might be differential risk associated with some variables when stratifying by whether or not cases were culture-confirmed, but this was not the case for any assessed risk factor included in meta-analysis. Regional differences may also contribute to heterogeneity in risk factors, although we did not find any differences based on epidemic or endemic setting in particular.In sum, we identified potential risk factors for symptomatic cholera infection ranging from environmental characteristics directly impacting exposure risk, to socioeconomic factors such as wealth or education, to intrinsic patient factors such as specific genetic features or gastric acidity. Many of these potential risk factors can and have been intervened upon in cholera control efforts , althougThe Journal of Infectious Diseases online. Consisting of data provided by the authors to benefit the reader, the posted materials are not copyedited and are the sole responsibility of the authors, so questions or comments should be addressed to the corresponding author.Supplementary materials are available at Supplementary Figure 1Click here for additional data file.Supplementary Figure 2Click here for additional data file.Supplementary Figure 3Click here for additional data file.Supplementary Figure 4Click here for additional data file.Supplementary Figure 5Click here for additional data file.Supplementary Figure 6Click here for additional data file.Supplementary Figure 7Click here for additional data file.Supplementary Figure 8Click here for additional data file.Supplementary Figure 9Click here for additional data file.Supplementary Figure 10Click here for additional data file.Supplementary Figure 11Click here for additional data file.Supplementary Figure 12Click here for additional data file.Supplementary Figure 13Click here for additional data file.Supplementary Figure 14Click here for additional data file.Supplementary MaterialClick here for additional data file.


Ovarian cancer is the primary cause of cancer-associated deaths among gynaecological malignancies. Increasing evidence suggests that microRNAs may be potential biomarkers for the diagnosis and prognosis of cancer. In this study, we conducted a systematic review and meta-analysis to summarize the global research and to evaluate the overall diagnostic accuracy of miRNAs in detecting ovarian cancer.A systematic literature search was conducted for relevant studies through July 20, 2017, in English databases , the Grey reference database and Chinese databases. Statistical analysis was conducted using OpenMetaAnalyst, STATA 14.0 and RevMan 5.3. Pooled sensitivity, specificity, and other parameters were used to assess the overall miRNA assay performance using a bivariate random-effects model (BRM). Meta-regression and subgroup analyses were performed to dissect the heterogeneity. Sensitivity analysis was performed to assess the robustness of our analysis, and the publication bias of the selected studies was assessed using Deeks’ funnel plot asymmetry test.Thirteen articles described 33 studies, including 1081 patients with ovarian cancer and 518 controls. The pooled results were as follows: sensitivity, 0.89 (95% CI: 0.84–0.93); specificity, 0.64 (95% CI: 0.56–0.72); positive likelihood ratio, 2.18 (95% CI: 1.89–2.51); negative likelihood ratio, 0.15 (95% CI: 0.11–0.22); and diagnostic odds ratio (DOR), 13.21 (95% CI: 9.00–19.38). We conducted subgroup analyses based on ethnicity, research design, and miRNA profiling and found that multiple miRNA panels were more accurate in detecting ovarian cancer, with a combined DOR of 30.06 (95% CI: 8.58–105.37).Per the meta-analysis, circulating miRNAs may be novel and non-invasive biomarkers for detecting ovarian cancer, particularly multiple miRNA panels, which have potential diagnostic value as screening tools in clinical practice.The online version of this article (10.1186/s13048-019-0482-8) contains supplementary material, which is available to authorized users. Per the International Agency for Cancer Research, Cancer Statistics 2016, ovarian cancer is the fifth most common cause of cancer-related deaths in the United States, with an estimated 22,280 new cases and 14,240 deaths in 2016 , and ovaMicroRNAs (miRNAs) are a class of small RNA molecules (19–22 nucleotides) that negatively regulate gene expression by inhibiting translation or degrading the target mRNA . MiRNA iCurrently, several groups have reported the potential diagnostic value of circulating miRNAs in various cancers such as oesophageal cancer, head and neck tumours, and gastric cancer –19; howeIn this study, we conducted a systematic review and meta-analysis to summarize global research and to evaluate the overall diagnostic accuracy of miRNAs in detecting ovarian cancer. We also explored an optimal miRNA and a potential panel to improve its diagnostic performance.thecochranelibrary.com), MEDLINE (through pubmed.gov), and EMBASE (through ovidsp.tx.ovid.com) and the Grey Reference database . We used the search strategy of subject headings and random words, which was designed by two researchers and included the following terms: (“miRNAs” OR “microRNAs” OR “miR*”) AND (“ovarian cancer” OR “ovary cancer” OR “ovarian neoplasms”) AND (“diagnos*” OR “ROC curve” OR “sensitivity” OR “specificity”). Search details are shown in Additional file The selected databases were searched through July 20, 2017, and included Chinese databases , English databases such as CENTRAL  studies were diagnostic and used circulating miRNAs; (2) histological subtype was specified as ovarian cancer; (3) sufficient data were reported to generate a 2 × 2 table for calculating true positives (TP), false positives (FP), false negatives (FN), and true negatives (TN); and (4) studies were not reviews, abstracts, animal studies or editorial articles. Studies were excluded for the reasons described in Fig.  studies Eligible articles were reviewed independently by two reviewers (DJK and CKW), and disagreements were resolved by consensus. The necessary information and data were extracted from the final eligible articles as follows: first author, year of publication, country of origin, subjects’ ethnicity, number of cases and controls, research types, median age, miRNA expression test methods, specimens, cut-off values, expression changes, and data needed for diagnostic meta-analyses (sensitivity and specificity) Table .Table 1CThe methodological qualities of the selected eligible articles were assessed by the Quality Assessment of Diagnostic Accuracy Studies 2 (QUADAS-2) score system. The QUADAS-2 tool combines the patient selection index, index test, reference standard, and flow and timing to evaluate risk of bias and applicability. The 7 items (four items on bias risk and three items on applicability) were assessed for all included articles. Two authors independently tested the pilot QUADAS-2 items (DJK and LZ), and discrepancies were resolved by a third author (XSW).2-based Cochran’s Q test was performed to assess the heterogeneity of the results across studies. If P < 0.10 and I2 was over 50%, it was considered heterogeneous, and meta-regression and subgroup analyses were performed to dissect the heterogeneity by the included studies’ characteristics. Sensitivity analysis was performed to assess the robustness of our analysis, and the publication bias of the selected studies was assessed using Deeks’ funnel plot asymmetry test.Data were analysed using OpenMetaAnalyst , Stata statistical software  and RevMan 5.3 . The pooled parameters sensitivity, specificity, positive likelihood ratio (PLR), negative likelihood ratio (NLR), and diagnostic odds ratio (DOR) and their 95% CIs were calculated to evaluate the overall diagnostic accuracy. A bivariate random effects model (BRM) was also used, and the correlation value was applied to evaluate threshold effects by BRM. In addition, a χThe flow diagram of the selected studies is depicted in Fig. Table 2 values of 98.49% (P < 0.001) for NLR, 86.79% (P < 0.001) for PLR and 74.16% (P < 0.001) for DOR, indicating significant heterogeneity; thus, the random-effects model was selected. The overall pooled results for sensitivity and specificity were 0.89 (95% CI: 0.84–0.93) and 0.64 (95% CI: 0.56–0.72), indicating moderate accuracy. The overall DOR, PLR, and NLR were 13.21 (95% CI: 9.00–19.38), 2.18 (95% CI: 1.89–2.51) and 0.15 (95% CI: 0.11–0.22), also indicating moderate accuracy.As shown in Figs. P < 0.05). We then conducted subgroup analyses to explore the potential diagnostic values of the miRNAs detected.Heterogeneity caused by differences in sensitivity and specificity is known as the threshold effect, which is an important source of heterogeneity in diagnostic tests. The correlation coefficient of logarithm sensitivity and specificity is a common approach to estimating a threshold effect. Our research data show that the correlation value was − 0.5001, revealing a threshold effect in this study. Moreover, we performed meta-regression to test the effect of ethnicity (Caucasian or Asian), study type (prospective or retrospective), and microRNA profiling  to explore heterogeneity sources . The results indicated that the prospective type DOR, 21.12) more accurately detected miRNA than the retrospective type  Fig. . We also1.12 moreP = 0.72)  Fig. b. The abOvarian cancer is the primary cause of cancer-associated death among gynaecological malignancies, representing 5% of all cancers in women . While pIn this meta-analysis, as a diagnostic biomarker, circulating miRNAs had a moderate diagnostic accuracy and achieved a combined DOR of 13.21, with a 0.89 pooled sensitivity and 0.64 pooled specificity, in discriminating ovarian cancer. As an advantage over a single indicator, the pooled DOR was 13.21, with higher values indicating better discriminatory test performance. We also found that the likelihood ratios (PLR and NLR) are more clinically meaningful for measuring diagnostic accuracy. In our meta-analysis, we found an NLR value of only 0.15, which refers to the probability of a person with ovarian cancer testing negative, divided by the probability of a person without ovarian cancer testing negative. These results revealed that circulating miRNAs can clearly distinguish ovarian cancer patients from healthy patients.As the study’s heterogeneity was large, we adopted the bivariate random effects model to reduce heterogeneity-related bias. Through the correlation value, we found heterogeneity to be a result of the threshold effect. The results from our meta-regression suggested that ethnicity, design type, and microRNA profiling might be sources of heterogeneity; therefore, subgroup analyses were used to investigate and explain the heterogeneity. The results suggested that circulating miRNAs could be used as ovarian cancer diagnostic biomarkers in both Caucasian and Asian populations with moderate diagnostic accuracy. Further, the Caucasian and Asian populations did not significantly differ. The diagnostic value of the research types varied considerably. Prospective studies were more reliable than retrospective studies and had more evidence with a relatively high diagnostic accuracy and a combined DOR of 21.12, suggesting that the miRNAs’ actual diagnostic value might be higher than that of the overall value. In addition, we conducted a subgroup analysis based on the number of miRNAs being profiled. The results suggested that multiple miRNA panels were more accurate with a combined DOR of 30.06 in detecting ovarian cancer. These results revealed that multiple miRNA panels more clearly distinguished ovarian cancer patients from healthy individuals, and more comprehensive prospective research may gain insight into the clinical value of miRNAs in the future.Compared with previous similar studies, comprehensive analysis of the diagnostic value of circulating miRNAs in ovarian cancer patients found that the pooled sensitivity was 0.75, specificity was 0.75, and the AUC was 0.82 . DiagnosAlthough this meta-analysis yielded promising results, several limitations must be addressed. First, the biological characteristics and mechanisms of the different miRNAs in ovarian cancer may differ, and only three miRNA markers  were repeated. In addition, methods for accurately and absolutely quantifying miRNAs are not uniformly normal, which limited applicability of the pooled analysis. Second, our data suggest that multiple miRNA panels can achieve better accuracy; however, the numbers required and the most efficient method for combining the miRNA panel remains unreported. Therefore, a rational and efficient miRNA panel should be investigated to improve future miRNA assay outcomes.Despite these limitations, our meta-analysis demonstrated that circulating miRNAs may be novel and non-invasive biomarkers for detecting ovarian cancer, especially multiple miRNA panels, which have potential diagnostic value as screening tools for clinical practice. Further large-scale prospective studies are warranted to improve the accuracy and explore the most effective miRNA combinations.Additional file 1:Additional Supporting Information may be found in the online version of this article. (DOCX 20 kb)


IntroductionInflammatory bowel disease (IBD) and its complications have been well-established. The literature shows an association between IBD and decreased bone mineral density in the adult population. However, most studies have reported an association between IBD and osteoporosis, while the risk of fractures has not been well-studied. The aim of this meta-analysis is to summarize the best available evidence regarding IBS and osteoporotic fractures.MethodsA review of the literature using the MEDLINE and EMBASE databases was performed during November 2017. We included cross-sectional and cohort studies that reported the relative risks, odds ratios, and hazard ratios comparing the risk of developing osteoporotic fractures among patients with IBD patients, both ulcerative colitis (UC) and Crohn’s disease (CD), versus patients without IBD as controls. The pooled odds ratio (OR) and 95% confidence interval (CI) were calculated using the generic inverse-variance method.Results2=42.3. No publication bias was found using the Egger regression test p=0.18. Sensitivity analysis showed that the inclusion of data on children by Kappelman et al. (2007) did not change the results.After a review of the literature, seven studies fulfilled the eligibility criteria established during the analysis. A significant association was found between IBD and osteoporosis, with a pooled OR of 1.32 . Low heterogeneity among the studies was found, IConclusionA significant association between IBD and the risk of developing osteoporotic fractures was observed in this study. There is a 32% increased risk, which is consistent with different cohort studies previously done. Inflammatory bowel disease (IBD), including ulcerative colitis (UC) and Crohn’s disease (CD), plays an important role as risk factors for the development of osteoporosis and osteoporotic fractures . The risWith the increased risk of developing osteoporosis, patients are at a greater risk of suffering osteoporotic fractures. IBD patients can have an increased risk of fractures by 40% as compared to the general population, which affects their quality of life and increases their morbidity .The mechanism of how IBD puts patients at an increased risk of developing osteoporosis and osteoporotic fractures is likely multifactorial. Often, steroids are prescribed to patients with IBD as treatment, and this can itself predispose patients to develop osteoporosis, however, the different studies that have been adjusted for glucocorticoid use have not reported a significant increase of osteoporosis in patients taking this medication . This beThe aim of this study is to summarize the best evidence correlating IBD patients with the risk of developing osteoporotic fractures versus control patients without IBD.Search strategyTwo investigators (DH and BB) personally reviewed two databases, MEDLINE and EMBASE, during the month of November 2017. The search strategy included terms and synonyms for “IBD,” “fractures,” and “osteoporosis.”Apart from the two major databases used for this meta-analysis, an additional manual review of the references of selected, included articles was also performed. This study meets the criteria checklist in accordance with the preferred reporting items for systematic reviews and meta-analysis (PRISMA) statement.Selection criteriaAny study, in order to be selected for this meta-analysis, had to fulfill the following parameters:- Case-control, cross-sectional, or cohort studies published as original articles in the two major databases used. These studies should investigate the risk of developing osteoporotic fractures in patients with IBD, either CD or UC.- Odds ratio (OR), relative risk (RR), or hazard ratio (HR) with 95% confidence intervals (CIs) or documentation and data enough to calculate the previous ratios were provided.In order to evaluate the quality of each study, the investigators independently used the validated Newcastle-Ottawa quality assessment scale Figure . This scData extractionA Microsoft Excel  data collection form was used to summarize the most relevant information obtained from these studies. This table contained the first author's last name, the country where the study was conducted, year of publication, demographics, the Newcastle-Ottawa quality assessment scale, the total number of participants, characteristics of the participants, the method used to diagnose IBD, the method used to determine fractures, adjusted effect estimates with 95% CI, and covariates that were adjusted in the multivariate analysis Table .To ensure accuracy, all investigators performed the data extraction process independently. Any data discrepancy was also resolved by referring back to the original articles.Statistical analysisData analysis was performed using Review Manager 5.3 software from the Cochrane Collaboration . Adjusted point estimates and standard errors from the individual study were combined using the generic inverse variance method of DerSimonian and Laird, which assigned the weight of each study based on its variance . In lighAn advanced search yielded 7237 articles on Medline and 2171 articles on EMBASE. After the exclusion of 6550 articles that were duplicated, 2858 underwent a title and abstract review. A total of 2802 articles were excluded, as they were case reports, book articles, letters to the editor, or review articles without the information needed for the analysis, leaving 56 articles for a full-length article review. A total of 45 articles were dismissed at this time, as they did not report the outcome we were looking for. Two articles did not have comparators so they were also excluded. A total of seven studies were used for statistical analysis , five ofThe overall analysis found a higher risk of osteoporotic fractures in patients with IBD, either UC or CD, as compared with the control individuals who did not have IBD. The odds ratio (OR) was 1.32 (95% CI: 1.20 - 1.45), P <0.001, as shown in Figure To measure heterogeneity among the studies, Cochran’s Q test and I2 were calculated. The I2 calculated for this study was 42.3, which represents low heterogeneity among the studies.The Egger regression test and funnel plots were used to assess publication bias . Egger'sThis study is a meta-analysis to evaluate the risk of developing osteoporotic fractures among IBD patients. After evaluating a total of seven studies  that met the inclusion criteria, with a total population of 242,524 patients, we found that patients with IBD have a 32% increased risk of developing osteoporotic fractures.This is the first meta-analysis that assesses the relationship between IBD and fractures. A lot of data has been obtained to assess the relationship between IBD and osteoporosis, which is defined as reduced mineral density. However, the risk for developing bone fractures is the one associated outcome that really has significant morbidity, mortality, and healthcare cost ,18.This meta-analysis takes into consideration research studies done all over the world, including the USA, UK, Hungary, China, and others. It collects more than 200,000 patients and records. We included a study done by Kappelman et al. 2009) that took into consideration the pediatric population  that too.We calculated the risk of fractures on IBD as a whole. We did not make any distinction between CD or UC, as some articles were just focused on CD while others included UC too. Further studies would be needed to evaluate the risk of fractures associated with the specific type of IBD.The data used for this meta-analysis were matched for the use of corticosteroids. Steroids by themselves put patients at increased risk of fractures. The most significant effect of steroids n the bones is inhibition of osteoblastic activity, which decreases bone formation. This inhibition in the number of osteoblasts is secondary not only to a decrease in the formation but also an increase in the death of mature osteoblasts. At the same time, glucocorticoids decrease the function of the remaining osteoblasts directly and indirectly through the inhibition of insulin-like growth factor I expression . Since mAfter adjusting for the use of this medication, the decrease in bone mass and the increase of developing bone fractures was significant enough to confirm that an inflammatory state is enough to make changes to the bone metabolism.A low I2 value on the statistical analysis suggests that the variation across studies is due to a low heterogeneity rather than chance.More studies should be done to calculate the risk to develop fractures in the pediatric population since early identification, risk factors modification, and treatment could slow the risk of bone mass loss and improve morbidity in that specific population.A significant association between IBD and the risk of developing osteoporotic fractures was observed in this study. There is a 32% increased risk of developing osteoporotic fractures in patients with IBD as compared to individuals without IBD used as controls. There was no significant increase of osteoporosis in IBD patients taking corticosteroids after this variable was adjusted through direct standardization.


Background: The level of interleukin-6 (IL-6) and its gene polymorphism are associated with the end-stage renal disease (ESRD) and the related complications. This study aimed to investigate the correction between IL-6 -174G/C polymorphism and ESRD by meta-analysis.Methods: Using the databases including PubMed, Embase, Cochrane library, CNKI, and CBM, the data of case-control studies on correlation between IL-6 -174G/C polymorphism and ESRD from database establishment to January 2016 were collected. According to inclusion and exclusion criteria, the quality of literatures was evaluated. The relevant research data were extracted, followed by meta-analysis using Revman 5.3 software . The combined odds ratio (OR) and 95% confidence interval (95%CI) of each genetic model were calculated, and the publication bias data was assessed using the Stata 12.0 software .Results: A total of five literatures were included, with 1199 cases in case group and 1089 cases in control group. Meta-analysis showed that, there was no significant correlation between each genetic model of IL-6 -174G/C polymorphism and ESRD . The race subgroup analysis showed that, there was no significant correlation between each genetic model of IL-6 -174G/C polymorphism and ESRD in the Caucasians (p > .05).Conclusion: IL-6 -174G/C polymorphism has no significant correlation with the susceptibility risk of ESRD, and may not be a risk factor for ESRD. With the development of disease, the chronic kidney disease will eventually progress to the end-stage renal disease (ESRD). Finally, the patients need to rely on renal replacement therapy or renal transplantation to sustain their lives.The inclusion criteria were as follows: (i) literatures about correlation between the IL-6 -174G/C polymorphism and the risk of ESRD; (ii) containing two groups of subjects, in which the subjects in case group was clinically diagnosed as ESRD, and the subjects in control group were the healthy population unrelated with ESRD; (iii) the data of IL6 -174G/C allele or genotype distribution could be obtained, and there were enough data to calculate the odds ratio (OR) and 95% confidence interval (95%CI); and (iv) the distribution of gene frequency in control group was in line with Hardy–Weinberg equilibrium (HWE).The exclusion criteria were as follows: (i) reviews, case reports, non-case-control studies; (ii) literatures in which the results did not contain the IL-6 -174G/C polymorphism; (iii) reports in which the genotypes were not complete, the full text or detailed data were not available, or there was no enough data to calculate OR or 95%CI; and (iv) if the data of was repeatedly published in a number of publications, the study of the highest quality was selected.Database of Pubmed, Embase, Cochrane library, CNKI, and CBM were retrieved, and the retrieval time ranged from database establishment to January 2016. The subject terms included “interleukin-6”, “IL-6”, “end-stage renal disease”, “ESRD” and “polymorphism”. The references attached to the retrieved literature were inquired by hand, aiming to find the literatures which were not included according to above inclusion criteria.According to the inclusion and exclusion criteria, two researchers independently conducted the literature screening, quality assessment and data extraction, with the cross checking. The disagreement between two researchers was solved by discussion or by the assistance of the third researcher. According to the evaluation criteria of case-control study in Newcastle Ottawa scale (NOS),χ2 test was used to determine whether the gene frequency distribution of the control group was consistent with HWE. p < .05 was considered not meeting HWE. Meta-analysis was carried out using Revman 5.3 software . The combined OR and 95%CI of five genetic models  were calculated. χ2 test was used to evaluate the heterogeneity in included literatures. p < .1 or I2 > 50% indicated the presence of heterogeneity among these literatures. The randomized effects model was used for analysis. Otherwise, the fixed effects model was applied. The subgroup analysis was performed based on the population race. The analysis of sensitivity was carried out by evaluating the stability after one-by-one eliminating study. Begg’s and Egger’s tests in Stata 12.0 software  were used to quantitatively evaluate the publication bias of the results of included literatures, and p < .05 indicated the publication bias.According to the literature retrieval strategy, 275 literatures were initially obtained. According to the inclusion and exclusion criteria, the repeated literatures, reviews, case reports, and non-case-control literatures were excluded. Seven case-control literaturesp = .38 , 95%CI , p = .68 (CC versus GG + CG (0.86): OR = 1.12, 95%CI, 1.46, p = .42 (CC); versus GG : , p = .58 : OR = 0.87, 95%CI , p = .50] : OR = 1.01, 95%CI , p = .50] .The sensitivity analysis was carried out by the method of evaluating the stability after one-by-one eliminating study eliminating single study. There was no significant change in the amount of the combined effects after in turn eliminating 1 literature. This indicated that the results of each literature were stable and reliable.p > .05 in both Begg’s and Egger’s test. This indicated that the included literature had no significant publication bias.As shown in ESRD patients are in the state of micro inflammation, which is closely related to vascular sclerosis, anemia, erythropoietin resistance, malnutrition, and infection. These are the important reasons for increased mortality rate of ESRD.,Most of the studies show that, the population carrying G allele in IL-6 -174G/C polymorphism have higher IL-6 level than carrying CC genotype. The G allele can promote the inflammatory response.The results of this study have significant heterogeneity, and the heterogeneity source can be found by subgroup analysis or sensitivity analysis. The subgroup analysis finds that, the heterogeneity is decreased in the studies of allele, recessive model and homozygous model in Caucasians, but the moderate heterogeneity still exists among the dominant model and heterozygous model. In this study, the difference in the characteristics of research subjects, disease diagnostic methods, alleles detection methods, region and environment can lead to the existence of heterogeneity. In this study, by using the one-by-one eliminating method, Ranganath et al.’s studyThis study has some limitations. Firstly, the sample size is relatively small. The race subgroup analysis is only conducted in the Caucasians and the subgroup races. The collected data are not comprehensive. Secondly, parts of the included literatures are lack of indexes such as age, sex, laboratory biochemical indicators in cases and control groups, so the stratified subgroup analysis based on these factors cannot be performed, and the combined results are affected by confounding factors. In addition, the results of this study are not consistent with the results of reported studies. The conclusion still needs to be further confirmed by more large-sample, multi-center, and high-quality case-control studies, for providing a theoretical basis for clarifying the function of IL-6 gene polymorphism in the pathogenesis of ESRD.


Inflammation may be a key pathophysiological mechanism in diabetic nephropathy (DN). Intercellular adhesion molecule 1 (ICAM1) is an acute phase marker of inflammation. ICAM1 rs5498 has been reported to be associated with the risk of DN. However, the previous findings were conflicting due to the limited sample sizes, different methodologies and ethnicities. Therefore, this study aimed to investigate the genetic association between ICAM1 rs5498 and the risk of DN.Two investigators independently searched the studies from the databases PubMed, Web of Science, the Cochrane Library, Chinese National Knowledge Infrastructure (CNKI) and Embase. Pooled odds ratios (ORs) with 95% confidence intervals (CIs) were used to assess the associations.p > 0.05). However, significant reduction of frequencies of the dominant model of ICAM1 rs5498 was only detected in the Caucasian subgroup  and type 1 diabetes mellitus subgroup .No significant association was detected between ICAM1 rs5498 and DN susceptibility in allelic and recessive models  contains supplementary material, which is available to authorized users. Diabetes mellitus (DM) is a metabolic disorder associated with chronic micro and macro vascular complications , 2. One Genome-wide scans have predicted that genes located at chromosome 19p13 might be susceptible to T1DM . The intGenetic association studies involving a common non-synonymous single nucleotide polymorphism (SNP) of ICAM1 (rs5498 E469K) in T1DM patients have been reported , 15. GujDue to limited sample sizes and inconsistent results of individual studies, this meta-analysis aimed to systematically evaluate the genetic association between rs5498 E469K and DN susceptibility.This meta-analysis followed the Cochrane Collaboration definition and PRISMA 2009 guidelines for meta-analysis and systematic review . To idenInclusion criteria: (1) case-control studies; (2) evaluating the association between ICAM1 rs5498 polymorphism and DN risk; (3) available genotype frequencies; (4) the genotype distribution in control groups was in the Hardy-Weinberg equilibrium (HWE). Exclusion criteria: (1) replicated studies; (2) studies in which the genotype or allele frequency could not be obtained; (3) data from meetings, case-reports, reviews or abstracts; (4) control group did not conform to HWE.Two researchers (Liya Liu and Dongling He) independently extracted the data from each eligible publication, and discussed to reach a consensus in case of disagreements. For each study, the following data were reviewed and collected: First author, year of publication, ethnicity, number of cases and controls, type of DM, mean age, gender, duration time, HbA1c, Body Mass Index (BMI), creatinine, High Density Lipoprotein (HDL), Systolic blood pressure (SBP), and Diastolic blood pressure (DBP).The study quality was independently assessed by Xiaojie Yan and Ling Fang. Any discrepancies in the assessment were resolved by the third author Xiaojie Yan. The quality of each included study was assessed according to the Newcastle–Ottawa Scale (NOS) . Otherwise, the random-effects model (the DerSimonian and Laird method) was used. Sensitivity analysis was performed to assess the effects of every individual study on the pooled results and the stability of results. Publication bias was assessed with funnel plots by Begg’s and Egger’s tests.Statistical analysis was conducted using the Stata software  and RevMan software (version 5.1). The pooled ORs were performed for null versus present genotype. The statistical significance of the pooled ORs under different genetic models  were determined by Z-test, and considered significant when A total of 357 studies were retrieved through the literature search. A total of 14 duplicate reports were excluded. After screening the titles, abstracts, and full-texts, seven eligible studies were included in the present meta-analysis and used for data extraction , 24, 25.p > 0.05)  , but not in T2DM subgroup . However, analyses stratified by sample size (sample size > 1000 and sample size < 1000) showed no significant association between allelic, dominant, additive and recessive models of ICAM1 and the risk of DN (p > 0.05)  of ICAM1 and increased risk of DN was detected (05) Fig. . Signifi05) Fig.  > 50% (F05) Fig.  > 50%  .The Begg’s and Egger’s tests were used to evaluate the publication bias. None of the funnel plots showed any evidence of publication bias Fig. . The p vThe present study demonstrated that the dominant model (GG + AG/AA) of ICAM1 decreased the risk of DN among Caucasian and T1DM subgroups. To the best of our knowledge, this is the first meta-analysis that assessed the genetic association between rs5498 and DN in DM patients. These findings are likely to provide a better and conclusive evaluation of ICAM1 rs5498 for the risk of DN.Diabetes mellitus (DM) is a heterogeneous deregulation of carbohydrate metabolism, characterized by chronic hyperglycemia that results from impaired glucose metabolism and the subequent increase in blood serum glucose concentration , 27. MulDiabetic nephropathy (DN) is one of the microvascular complications in DM and a main cause of end-stage renal disease , 36. DN In the present meta-analysis, we included studies of both T1DM and T2DM patients with and without DN. A significant association was observed between the dominant model of ICAM1 and DN in T1DM, but not in T2DM subgroup, which may indicate that ICAM1 rs5498 was a susceptible factor in T1DM as well as DN. Several groups have reported genetic association studies of ICAM1 rs5498 in T1DM. Guja et al. reported that the transmission of the G allele of SNP K469E (A/G) is increased in Romanian T1DM families . NishimuNotably, significant association was only detected in the Caucasian subgroup, but not in the Asian subgroup. Previous studies indicated that this polymorphism was found to be associated with adult-onset T1DM in Japanese patients , but not2 ranged from 62 to 95). Significant heterogeneity was also found in Asians, sample size < 1000, and T2D subgroups in all genotype models (I2 ranged from 67 to 97). There could be several reasons for these heterogeneities. Firstly, most of these studies had small sample size. Secondly, different strategies were applied in various studies, including control source, population stratification, phenotype selection, genotyping methods, etc.Statistically significant heterogeneity across studies was observed in all genotype models  for quality assessment of included studies. (DOC 32 kb)


Although several studies have proved the relationship between the prognostic value of miRNA-15a and different types of cancer, the result remains controversial. Thus, a meta-analysis was conducted to clarify the prognostic value of miRNA-15a expression level in human cancers. We enrolled appropriate literature by searching the databases of PubMed, Embase, and Web of Science. Subsequently, we extracted HRs and their 95% CIs and calculated pooled results of miRNA-15a for overall survival (OS) and disease-free survival (DFS). Besides, subgroup analysis, sensitivity analysis, and publication bias were also revealed in this study. We also further validated this meta-analysis using the Kaplan-Meier plotter database. 10 studies, including 1616 patients, were embraced in our meta-analysis. The result showed the lower expression of miRNA-15a significantly predicted adverse OS , but there is no significant association between the expressing level and DFS in cancer patient . Based on Kaplan-Meier plotter database, we found the same results in bladder Carcinoma, head-neck squamous cell carcinoma, liver hepatocellular carcinoma, lung squamous cell carcinoma, pancreatic ductal adenocarcinoma, rectum adenocarcinoma, stomach adenocarcinoma, and uterine corpus endometrial carcinoma, but opposite results were found in cervical squamous cell carcinoma and esophageal carcinoma. Low expressing levels of miRNA-15a indicated poor OS, while miRNA-15a can be used as a prediction biomarker in different cancer types. MicroRNAs (miRNAs), endogenous noncoding single‐stranded RNAs with 20-25 nucleotides in length, regulate gene expression by binding with imperfect complementarity to the 3late gene expression at specific mRNAs. They may regulate approximately 60% genes in human protein-coding , 2. PrevMiRNA-15a is an important part of the miRNA-15 family , which hThe meta-analysis according to the guidelines of the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) statement and methods.  We retriEach study was revised by two reviewers (FRY and TTL) followed a standard data extraction form. Discrepancies were solved by fully discussing with HJL or further reviewed by XRL. Eligible studies must comply with the following criteria: (1) study about the correlation of miRNA-15a with any type of cancer prognosis; (2) publication details: disease name, publication year, and first author's last name; (3) the effect size being evaluated using multivariate HRs with 95% confidence intervals (CI) for OS or DFS, if the HRs cannot be obtained directly from the original studies, data were calculated by Kaplan-Meier curves according to the method provided by Tierney et al.  Studies All the enrolled literature was evaluated by Newcastle-Ottawa scale. This quality evaluation scale of literature contains the following three aspects: study groups selection, comparison of study groups, and measuring of outcomes. The score of this scale is ranged from 0 to 9.http://kmplot.com/analysis/index.php?p=service&cancer=pancancer_mirna). [ P-value is less than 0.05, it is considered statistically significant.To further validate and complement this meta-analysis, we used the Kaplan-Meier plotter database and miRpower for pan-cancer tool to analyze the relationship between miRNA-15a expression and OS of human cancers (_mirna).  If the P I2 statistic. We preferred to use fixed-model if there is no significant heterogeneity (P>0.1 and I2<50%); otherwise, we used the random-effect model. Publication bias was evaluated using the funnel plot with Begg's statistical test. We use the STATA software for statistical analysis .The pooled HRs were calculated by HRs with 95% CIs. Heterogeneity between pooled studies was evaluated by Cochran's Q test and Higgin'sWe used the search strategy provided above to retrieve 488 records from online databases. By manually reviewing titles and abstracts, we rejected duplicates and unavailable literature. Then 82 articles were selected for full-text browsing. Finally, ten studies were included and there was no additional article by consulting the cross-references. Ten studies consisted of 1616 samples from China, US, Australia, Greece, Japan, and Thailand and were included to evaluate the relationship between miRNA-15a expression and cancer prognosis. Eight of the 10 studies detected the expression of miRNA-15a in tumor tissue and another two studies , 23 deteAfter using Newcastle–Ottawa scale to assess the quality of studies, the study quality score >6 was included. The main characteristics of the eligible studies are summarized in I2 =66.6%, P=0.001), we chose a random-effect model to compare between low expression of miRNA-15a and high expression. The result showed that lower expression of miRNA-15a indicating adverse OS . We found that there is no significant association between the expressing level and DFS in cancer patients , serum or bone marrow specimen , and Asia patients . The association between downregulation of micro-RNA15a expression and shorter OS time was not obvious in the subgroup of European and colorectal cancer patients.  P=0.150). However, in the survival analysis of a single cancer types, we found that low-expression miRNA-15a was significantly associated with worse OS in bladder Carcinoma , head-neck squamous cell carcinoma , liver hepatocellular carcinoma , lung squamous cell carcinoma , pancreatic ductal adenocarcinoma , rectum adenocarcinoma , stomach adenocarcinoma , and uterine corpus endometrial carcinoma , whereas the results were opposite in cervical squamous cell carcinoma  and esophageal carcinoma   . In the To assess the stability of the meta-analysis results, we carried out sensitivity analysis by excluding studies one by one. After each study is excluded, the pooled HR was recalculated. The result showed that the stability of the entire study was not affected by a certain study .Publication bias for the OS meta-analysis among included studies was evaluated by Bgger's funnel plot . The funAs a class of tiny regulatory RNA molecules, miRNA plays an important role in gene expression and diverse biological processes. They can blind to the 3′UTR and 5′UTR of their target mRNAs. A large number of studies have demonstrated that tumor invasion and metastasis are regulated by miRNAs, and they are used to analyze the prognosis of a variety of tumors. For example, Wang et al. have demonstrated that overexpression of miRNA-34a significantly predicted good OS in cancer patients.  Zhang etMiRNA-15a was the first miRNA to be discovered as a tumor suppressor in CLL.  MiRNA-15I2=70.6, P=0.017). When we removed the study of Kontos et al. from the subgroup of colorectal cancer types, no significant heterogeneity was found. In Kontos et al.'s study, low expression group significantly reduced risk of death, which was contrary to the other three studies, but it was not statistically significant . Although there are significant statistical differences in the analysis of DFS , OS is still the best indicator for prognosis in clinical practice. Therefore, we concluded that the main source of heterogeneity is the study by Kontos et al. In addition, we performed survival analysis of human cancers through the Kaplan–Meier plotter database and found that the results of esophageal carcinoma and breast cancer were contrary to our meta-analysis; we analyze the reasons that lead to this result which may have the following aspects: (1) the sample size is too small, for esophageal carcinoma, only 184 cases in the database and 106 cases in the meta-analysis. For breast cancer, comparing 1076 cases in the database, there is only one literature and 230 cases in the meta-analysis. (2) Pathological classification of esophageal cancer is not given in the database, which may also be the cause of inconsistent results. Interestingly, we found in the database that lower expression of miRNA-15a significantly predicted adverse OS in CRC, which is inconsistent with the result of our meta-analysis. However, considering the survival analysis of CRC is still controversial in our meta-analysis, so there should be more clinical studies in the future to discover the relationship between miRNA-15a and the prognosis of CRC. All in all, through our meta-analysis, we found that miRNA-15a may be an important biomarker for predicting the clinical outcome of cancer patients.Recently, a large number of reports indicate that there is a correlation between the expression of miRNA-15a and the prognosis of various kinds of cancers. To identify the prognostic value of miRNA-15a, we performed this meta-analysis of 10 studies and 1616 patients with 7 cancers. We found that lower expression of miRNA-15a was associated with shorter OS. In other words, patients with higher expression of miRNA-15a have longer OS time than low expression levels. The expression levels of miRNA-15a are downregulated in most cancer types, and they may function as tumor suppressor genes. But obvious heterogeneity was found in our meta-analysis, so we conducted a subgroup analysis by cancer types, race, publication year, and material of miRNA-15a. In the subgroup of cancer types, we found significant heterogeneity in CRC group (To the best of our knowledge, there was no previous meta-analysis of the relationship between miRNA-15a expression levels and the prognosis of cancer patients. Nevertheless, our research has some limitations. Firstly, only 10 studies with 1616 patients were included in our analysis; this may make the result inaccurate, and the reason for the heterogeneity may be because the sample size is too small. Secondly, the design of studies, cut-off value, and measure methods were distinct in different researches, and these factors may have a certain impact on the results of our analysis. Third, we cannot find the HRs and their 95% CIs directly from some literature; thus, we had to estimate results from the Kaplan-Meier curve. This reduced the credibility of our results. Fourth, histopathology is the gold standard for tumor diagnosis, because of its high accuracy. In clinical applications, blood samples were easier to obtain than tissues. Therefore, in the future researches, increasing blood sample applications can provide evidence for clinical diagnosis and treatment in the absence of tissue samples case.As shown in our meta-analysis and bioinformatics, low expression of miRNA-15a may indicate poor OS in cancer patients. But there is no significant association between the expression level and DFS in patient with cancer and there is also controversy in esophageal cancer, breast cancer, and colorectal cancer. Further clinical studies are needed to demonstrate the association between miRNA-15a and cancer prognosis as well as treatment efficiency.


METH gene polymorphisms and prostate cancer susceptibility remains ambiguous. Thus, we performed an updated meta-analysis of METH single-nucleotide polymorphism rs1805087 involving 12 independent case-control studies comprising 9986 prostate cancer patients and 40134 controls. The odds ratio and 95% confidence intervals were applied to evaluate the relation of this single-nucleotide polymorphism with prostate cancer. Statistical analysis was performed in STATA 11.0. A significant association was found between rs1805087 and increased prostate cancer risk, overall and with Hardy–Weinberg equilibrium. In subgroup analyses , similar associations were observed . Thus, the common polymorphism (rs1805087) of METH may be associated with increased prostate cancer risk. Further studies with a larger sample size and detailed gene–environment interactions should be conducted to identify the role of METH polymorphisms in prostate cancer susceptibility.Methionine synthase  is a key enzyme in the folate pathway, which plays a critical role in the synthesis, repair, and methylation of DNA. The association between  MLH1, MSH2, MSH6, PMS2, or EPCAM) increases the risk of PCa two- to fivefold, in comparison with the general population [Prostate cancer (PCa) is a leading cause of cancer-associated deaths among men in developed countries . Althougpulation ,6–8.METH has been found at position 2756  [TP53 [Folate metabolism–associated genes involved in both DNA methylation and repair are thought to play an important part in tumorigenesis . The gen1805087) . This po1805087) . Moreove7) [TP53 .METH rs1805087 single-nucleotide polymorphism (SNP) on the risk of PCa have yielded inconsistent results. Collin et al. (2009) [MTHFR rs1801133, MTHFR rs1801131, METH rs1805087, METHR rs1801394, MTHFD1 rs2236225, SLC19A1/RFC1 rs1051266, SHMT1 rs1979277, and FOLH1 rs202676, on PCa risk in a meta-analysis and found no significant effects of any of these SNPs on PCa susceptibility. The study included eight case-control studies (five of which have not been published) on rs1805087. Since 2009, many more original research articles have been published, making it necessary to combine the data from all the aforementioned studies and reanalyze the association between rs1805087 and PCa susceptibility.Some studies that have addressed the effects of the . (2009)  investigThe objective of our study was to examine associations between rs1805087 genotypes and the risk of PCa in larger samples by meta-analysis (9986 PCa patients and 40134 controls from 12 studies) –27.Using different combinations of key terms, 24 studies were identified in PubMed and Embase. As shown in https://www.ncbi.nlm.nih.gov/projects/SNP/snp_ref.cgi?rs=1805087). The various genotyping methods employed in the analyzed studies included polymerase chain reaction with restriction fragment length polymorphism (PCR-FLIP), sequencing, TaqMan, GeneChip, Illumina, and SNaPshot analyses [Finally, we identified 12 case-control studies –27, invoanalyses –27.Pheterogeneity = 0.263, P = 0.048; dominant model: OR = 1.20, 95% CI 1.05–1.38, Pheterogeneity = 0.000, P = 0.009; allelic comparison: OR = 1.15, 95% CI 1.04–1.29, Pheterogeneity = 0.000, P = 0.008; Pheterogeneity = 0.000, P = 0.012; The overall meta-analysis results suggested significant associations between two rs1805087 genotypes and increased PCa susceptibility for all genetic models  but not among population-based controls . A growing number of original research articles and unpublished studies makes it necessary to combine data from all the studies to date and to reanalyze the association between rs1805087 and PCa risk.Meta-analysis is a means of increasing the effective sample size by pooling data from association studies, thereby enhancing the statistical power of estimation of genetic effects . In 2009In this study, we analyzed the associations between genotypes of rs1805087 and PCa risk by the meta-analysis method to reach a statistically backed conclusion. The potential ability of rs1805087 to increase PCa risk was investigated in a sample of 9986 cases and 40134 controls from seven published and five unpublished case-control studies. The conclusions from our results are 1) individuals who carry the rs1805087 G allele have increased susceptibility to PCa; 2) rs1805087 may increase PCa risk among both Asians and whites. These conclusions are reported for the first time and warrant continuation of research into rs1805087 regarding other types of cancer.Of the 12 studies, excluding the five unpublished studies ,19,22,26Although we made considerable efforts and devoted substantial resources to testing the possible association between rs1805087 and PCa risk, there are some limitations inherent in the included studies. First, despite inclusion of all the eligible studies, the resultant sample size is still not large enough ,27; thisMETH SNP rs1805087 may increase susceptibility to PCa. Further well-designed studies with a large sample size, different ethnicities, and detailed environmental factors are needed to validate the conclusions of our meta-analysis.In summary, our meta-analysis shows that We searched the PubMed and Embase databases for all articles on the association between rs1805087 and PCa risk up to January 20, 2018. The medical subject headings and key words used for the search were “MTR or methionine synthase,” and “prostate cancer or carcinoma,” and “polymorphism or variant.” The electronic search also covered the reference lists of all the identified articles and reviews to find possible additional original reports. All the included studies met the following criteria: (1) the design was “case-control study,” (2) the association between rs1805087 and PCa risk was explored (both published and unpublished), and (3) carcinoma cases were diagnosed histopathologically. The major exclusion criteria were (1) duplicate data; (2) abstract only, commentary, review, or editorial publications; and (3) insufficient reporting of data.The following information was collected: the first author’s last name, year of publication, country of origin, ethnicity, source of the control  and the Hardy–Weinberg equilibrium (HWE) of the control, total number of cases/controls, and genotyping method. Subgroups were classified based on the source of the control, on ethnicity, genotyping method, and publication status.The Newcastle-Ottawa Scale (NOS)  was sele2-based Q statistic and was considered statistically significant at P < 0.10 [P > 0.10, the pooled OR of each study was calculated via the fixed-effects model . Otherwise, the random-effects model (the DerSimonian and Laird method) was used [Z-test, and data with P < 0.05 were considered statistically significant. The departure of frequencies of rs1805087 from the expected values under the HWE was assessed by the χ2 test among controls; P < 0.05 was assumed to denote a significant disequilibrium. Publication bias was detected with Egger’s linear regression method and funnel plot; the funnel plot asymmetry was assessed by Begg’s test; data with P < 0.05 were considered statistically significant [http://biostat.mc.vanderbilt.edu/wiki/Main/PowerSampleSize#Windows) [The strength of the association between rs1805087 and PCa risk was measured by odds ratios (ORs) with 95% confidence intervals (CIs). Pooled ORs were obtained from a combination of studies by heterozygote comparison, homozygote comparison, dominant and recessive models, and allelic comparison. Heterogeneity among the studies was checked using the χP < 0.10 . At P > was used ,44. The nificant . All staWindows) .PRIZMA 2009 Checklist


MTHFR (methylenetetrahydrofolate reductase) gene and colorectal polyp risk. This discrepancy led us to assess the genetic impact of the two polymorphisms on the susceptibility to colorectal polyps.There are several studies with inconsistent conclusions regarding the association between the rs1801133 and rs1801131 polymorphisms within the MTHFR rs1801133 and rs1801131 polymorphisms. The P value of association test, RRs (risk ratios) and 95% CIs  were mainly produced.A meta-analysis was carried out for quantitative synthesis. According to the inclusion/exclusion criteria, we retrieved, screened and selected all published articles related to colorectal polyps and the P values in the test of association were larger than 0.05. Nevertheless, pooling results in the “UK” subgroup of rs1801131, comprising five studies (1257 cases/1407 controls), indicated an elevated risk in colorectal polyp cases in comparison with controls, under the genetic models of CC vs. AA  and CC vs. AA+AC .A total of twenty-three case-control studies were included from twenty-two eligible articles. Pooling the results of both rs1801133 and rs1801131 polymorphisms in the overall population suggested a nonsignificant association between colorectal polyp cases and controls, in that all MTHFR rs1801131 is more likely to be a genetic risk factor for colorectal polyps in the UK region, although this finding should be verified with a larger sample size.The C/C genotype of The online version of this article (10.1186/s12881-019-0822-y) contains supplementary material, which is available to authorized users. Colorectal polyps exhibit different morphologic features with flat, depressed, serrated, sessile or pedunculated shapes and are often regarded as benign protrusions of the colon and rectum mucosa , 2. TherMTHFR) gene is essential for the folate cycle and homocysteine metabolism [MTHFR gene [MTHFR rs1801133 and rs1801131 polymorphisms were reportedly associated with an enhanced risk of colorectal adenomatous polyp patients in the Korean population [MTHFR rs1801133 polymorphism and colorectal adenomatous polyp susceptibility was reported in the Dutch [The 5,10-methylenetertahydrofolate reductase  gathered the relative records through searching the databases, namely, PubMed, WOS (Web of Science), and EMBASE (Excerpta Medica Database), prior to March 2018. The PRISMA  guidelines were followed . The seaP value of HWE (Hardy-Weinberg Equilibrium). The included studies should provide the genotype frequency data of the control group, which also must be in line with the requirement of HWE. We summarized the main features of the included studies, such as first author name, publication year, polymorphism genotype frequency, country, ethnicity, genotyping assay, and P value of HWE. We also utilized quality assessment  to determine the quality score of the enrolled studies. Studies with poor quality (NOS score less than five) were excluded.We carefully extracted the data from the above selected studies. The chi-squared test was applied for the calculation of the Passociation, risk ratios (RRs) and 95% confidence intervals (CIs) through the association test. The Pheterogeneity value of Cochran’s Q statistic > 0.1 or I2 value < 50% led us to use a fixed-effects model. Six genetic models were used: allele T vs. allele C for rs1801133, allele C vs. allele A for rs1801131 ; TT vs. CC, CC vs. AA (homozygote); CT vs. CC, AC vs. AA (heterozygote); CT + TT vs. CC, AC + CC vs. AA (dominant); TT vs. CC + CT, CC vs. AA+AC (recessive); carrier T vs. carrier C, carrier C vs. carrier A (carrier).We obtained the We also carried out a sensitivity analysis and subgroup analyses for all genetic models to evaluate the data stability and source of heterogeneity. Briefly, we omitted each included study in turn to acquire a group of meta-analysis estimations. The omitted study was regarded as the probable heterogeneity source if we detected an obvious alteration of RR and 95% CI value. Subgroup analyses were also carried out, taking the factors of country, ethnicity (Caucasian/Asian) and disease type (hyperplastic polyps/ adenomatous polyps) into consideration.P values of Begg’s test and Egger’s test were > 0.05. STATA/SE software  was utilized for all the above tests.We conducted both Begg’s test (Begg’s funnel plot) and Egger’s test (Egger’s publication bias plot) to evaluate possible publication bias. The absence of a large publication bias was considered when the n = 22), WOS (n = 83), and EMBASE (n = 48). After excluding duplicate records, a total of 115 records were filtered by our criteria. The following 88 records were excluded: reviews (n = 31), mouse data (n = 4), case reports or trials (n = 7), meta-analyses (n = 6), meeting or conference abstracts (n = 8); other genes (n = 9), non-SNP or nonpolyp data (n = 23). Subsequently, twenty-seven full-text articles were evaluated for eligibility. Five articles lacked control or T/T genotype data. Finally, a total of twenty-two articles [MTHFR rs1801133, and ten case-control studies were included for that of MTHFR rs1801131. In addition, one study in which the TT genotype frequency of case and control groups for rs1801133 equaled zero was not included in the meta-analysis under the TT vs. CC (homozygote) and TT vs. CC + CT (recessive) models. The PRISMA-based analysis flowchart is shown in Fig. We initially identified a total of 153 records by searching three databases, namely, PubMed ; TT vs. CC (P = 0.454); CT vs. CC (P = 0.077); CT + TT vs. CC (P = 0.079); TT vs. CC + CT (P = 0.847); carrier T vs. carrier C (P = 0.322). We also conducted subgroup analyses by country, ethnicity (Caucasian/Asian) and disease type (hyperplastic polyps/adenomatous polyps). A similar nonsignificant genetic relationship was observed for all the models ; TT vs. CC (P = 0.641); CT vs. CC (P = 0.351); CT + TT vs. CC (P = 0.511); TT vs. CC + CT (P = 0.436); or carrier T vs. carrier C (P = 0.831). In the subgroup analysis of “adenomatous polyps”, we also did not observe a statistically significant association under the allele T vs. allele C (Table P = 0.153); TT vs. CC (P = 0.377); CT vs. CC (P = 0.113); CT + TT vs. CC (P = 0.098); TT vs. CC + CT (P = 0.696); and carrier T vs. carrier C (P = 0.331). We show the forest plots of the subgroup analyses based on disease type under the allele T vs. allele C model in Fig. MTHFR rs1801133 does not appear to be significantly linked to susceptibility to colorectal polyps.First, we carried out a meta-analysis to investigate the genetic relationship between MTHFR rs1801131. Pooled analysis in the overall population (Table P > 0.05). The results of the subgroup analysis for the UK, containing five studies of 1257 cases/1407 controls, suggested an increased risk in cases of colorectal polyps compared with controls under the genetic models of CC vs. AA  and CC vs. AA+AC . We showed the related forest plots in Figs. P > 0.05, Table P = 0.138); CC vs. AA (P = 0.114); AC vs. AA (P = 0.576); AC + CC vs. AA (P = 0.303); CC vs. AA+AC (P = 0.122); or carrier T vs. carrier C (P = 0.376). Thus, the C/C genotype of the MTHFR rs1801131 polymorphism may be related to an enhanced colorectal polyp risk in the UK population.Next, ten studies containing 2951 cases and 3527 controls were included in the meta-analysis of 2 < 50.0%, P value of heterogeneity > 0.1). Thus, a fixed-effects model was applied. We also conducted both Begg’s test and Egger’s test to assess the presence of publication bias. As shown in Table P values of Begg’s test and Egger’s test were larger than 0.05 in all genetic models, indicating the absence of large publication bias. We showed Begg’s funnel plot and the association between the MTHFR rs1801131 polymorphism and colorectal polyp risk under the CC vs. AA model in Fig. MTHFR rs1801131; other data not shown), suggesting the reliability of pooling outcomes.In addition, we evaluated the between-study heterogeneity and did not detect remarkable heterogeneity in any of the above comparisons  and adenoma. For example, in 2005, Kono, S. and colleague included a total of 16 case-control studies for a meta-analysis on the genetic relationship between MTHFR rs1801133 polymorphism and the risk of colorectal cancer and reported the potential role of the TT genotype in reduced CRC susceptibility [MTHFR rs1801133 and rs1801131 polymorphisms may confer reduced susceptibility to CRC patients [MTHFR rs1801133 polymorphism and overall cancer susceptibility and found that the TT genotype of MTHFR rs1801133 was associated with a decreased risk in CRC patients with lifelong hyperhomocysteinemia and hence hypomethylation [MTHFR rs1801133 polymorphism was related to a reduced risk of CRC, particularly in the Asian population [MTHFR polymorphism, especially rs1801133, on CRC risk. However, inconsistent results regarding the role of the MTHFR polymorphism in the risk of colorectal adenoma were observed in the quantitative synthesis.Several meta-analyses have reported the role of pulation . These dMTHFR rs1801133 and rs1801131 polymorphisms may have no increasing or decreasing effect on the risk of colorectal adenoma patients [MTHFR polymorphisms did not reach a genome-wide significant P value [MTHFR rs1801133 polymorphism may be associated with high susceptibility to colorectal adenoma in patients with poor folate status [the MTHFR rs1801133 polymorphism, but with a relatively lower statistical power [Meta-analysis of Huang, Y. et al. revealed that patients . In addi P value . Howeveral power .MTHFR rs1801133 polymorphism in the susceptibility to colorectal adenomas in terms of colorectal polyps by means of a meta-analysis containing twenty-three case-control studies with 8339 cases and 17,731 controls. Our findings did not show any association between the MTHFR rs1801133 polymorphism and the risk of colorectal adenomatous polyps or hyperplastic polyps.In this study, we intended to reassess the role of the MTHFR rs1801131 polymorphism has a significant influence on an increased risk of colorectal polyps in the UK population. The A to C substitution in exon seven of MTHFR gene-induced abnormal enzymatic activity, homocysteine or folate level and DNA methylation/synthesis may be implicated in this process. It is noteworthy that, based on the requirement of meta-analysis for the enrolled case-control number, we evaluated only the subgroup analysis data with at least three case-control studies. Therefore, the subgroup analysis data for Australia, the USA, Korea, and Japan, with one or two case-control studies, exhibits very limited statistical power. We still cannot exclude the potential effect of the MTHFR rs1801131 polymorphism in colorectal polyp patients of other regions.Moreover, we performed another meta-analysis of ten case-control studies with 2969 cases and 3527 controls and found that the C/C genotype of the MTHFR rs1801131 polymorphism and hyperplastic polyp risk [MTHFR rs1801133 and rs1801131 polymorphisms has been suggested in the susceptibility to colorectal cancer [MTHFR variants in the malignant conversion from colorectal polyp.The case-control studies in our analysis were screened by fulfilling our strict selection criteria. All the studies exhibit high quality. In addition, we observed no heterogeneity in any of the Mantel-Haenszel statistics and excluded the large publication bias. Moreover, the stability of the statistical outcomes was detected by the sensitivity analysis. Nevertheless, we are also aware of several limitations. The main problem is the small sample size in the included case-control studies. For example, only one case-control study analyzed the correlation between the lyp risk . Second,l cancer , additioMTHFR rs1801131, rather than rs1801133, is more likely to be associated with an increased susceptibility to colorectal polyps in the UK population. Additionally, the C/C genotype of MTHFR rs1801131 may confer an increased susceptibility to patients with colorectal polyps in the UK region. However, this conclusion merits further confirmation with a larger sample size.Taken together, our findings conclude that Additional file 1:Table S1. The search terms used with the PubMed, WOS and EMBASE databases. (DOCX 30 kb)