Only a minority of individuals who undergo cancer genetic counseling experience heightened levels of psychological distress, but many more experience a range of cancer genetic-specific psychosocial problems. The aim of this study was to estimate the prevalence of such psychosocial problems, and to identify possible demographic and clinical variables associated significantly with them. Consenting individuals scheduled to undergo cancer genetic counseling completed the Psychosocial Aspects of Hereditary Cancer (PAHC) questionnaire, the Hospital Anxiety and Depression Scale (HADS) and the Distress Thermometer (DT) prior to or immediately following their counseling session. More than half of the 137 participants reported problems on three or more domains of the PAHC, most often in the domains ‘living with cancer’ (84 %), ‘family issues’ (46 %), ‘hereditary predisposition’ (45 %), and ‘child-related issues’ (42 %). Correlations between the PAHC, the HADS and the DT were low. Previous contact with a psychosocial worker, and having a personal history of cancer were associated significantly with HADS scores, but explained little variance (9 %). No background variables were associated significantly with the DT. Previous contact with a psychosocial worker, and having children were significantly associated with several PAHC domains, again explaining only a small percentage of the variance (2–14 %). The majority of counselees experience specific cancer genetic counseling-related psychosocial problems. Only a few background variables are associated significantly with distress or psychosocial problems. Thus we recommend using the PAHC or a similar problem-oriented questionnaire routinely in cancer genetic counseling to identify individuals with such problems. One of the main messages of studies on the psychosocial impact of genetic counseling for cancer is that, after the process of genetic counseling and risk assessment has been completed, distress levels for the majority of counselees return to or are even lower than baseline levels –3. HowevThe psychosocial impact of genetic counseling is most frequently measured with the Hospital Anxiety and Depression scale (HADS), the State Trait Anxiety Inventory, the Impact of Event Scale, or the Center for Epidemiological Studies Depression Scale –7. HowevSeveral methods are available to assist genetic counselors in detecting counselees with serious psychosocial problems. It has been proposed to use sociodemographic and clinical risk factors to identify  distressed individuals , 12. VadIncreasingly, the Distress Thermometer (DT) with an accompanying problem checklist is being recommended as a first line screening method for distress in daily clinical oncology practice . The DT,Recently, we developed the Psychosocial Aspects of Hereditary Cancer (PAHC) questionnaire as a tool for identifying psychosocial issues and concerns experienced during cancer genetic counseling . The PAHKnowledge of the specific psychosocial problems and distress levels experienced by counselees, as well as factors that may be associated with such problems can provide genetic counselors with useful information that they can use during the genetic counseling session. In this paper, we report on a study of the prevalence of cancer genetic counseling-specific psychosocial problems and their association with more generalized distress as assessed by the HADS and the DT. We also investigated whether sociodemographic and clinical variables are associated significantly with psychosocial problems and psychological distress experienced during cancer genetic counseling.The data reported here were collected as a part of a larger study that evaluated the screening properties of the PAHC questionnaire and the DT in the cancer genetic counseling setting . The insIndividuals were eligible to participate when they were scheduled for a visit at the family cancer clinic of The Netherlands Cancer Institute to undergo genetic counseling for any type of hereditary cancer syndrome in the period January–December, 2010, were over 18 years of age, and had a sufficient command of the Dutch language.Eligible counselees received a letter of invitation from the head of the family cancer clinic and, if interested, were requested to return a signed consent form by mail. A reminder letter was sent 1 week before the genetic counseling session. Participants completed a questionnaire on a touchscreen computer at the clinic with demographic questions, the PAHC questionnaire, the DT and the HADS. The preference was to have the questionnaire completed prior to the counseling, but this was not always feasible due to planning issues. Thus counselees completed the questionnaire immediately prior to their scheduled genetic counseling session or immediately thereafter.The counselees’ age, sex, marital status, education level, number of children, the number of affected first degree relatives, and use of psychosocial services in the past for any problem  were obtained via self-report. Data on whether (s)he was diagnosed with cancer in the past and, if so, at what age, and whether there was a known gene mutation in the family were extracted from the medical records.The PAHC questionnaire consists of 26 questions addressing psychosocial problems and concerns that are specifically relevant to counselees within the cancer genetic counseling and testing setting. The content of the PAHC questionnaire is organized into the following six domains: (1) hereditary predisposition; (2) practical issues; (3) family and social issues; (4) general emotions; (5) living with cancer; and, for those who have children (6) children-related issues. The number of items per domain varies between 2 and 6. All 26 items are scored on a 4-point, Likert-type scale ranging from 1  to 4 (“very much”). Based on a detailed analysis of the screening properties of the PAHC questionnaire, a threshold was established for clinical relevance . SpecifiThe HADS was used to assess general psychological distress. It includes 14 questions and yields a total score and subscale scores for anxiety and depression. In the current analysis, we used only the total score, with a possible range of 0–42. Higher scores represent higher levels of distress. The HADS has been validated for use in the Netherlands .t tests were used to examine potential differences in responses to the PAHC questionnaire, the HADS and the DT as a function of timing of questionnaire completion . The association between the PAHC questionnaire domains, the HADS and the DT was assessed by calculating Pearson’s correlation coefficients and partial correlations that controlled for inter-correlations between the domains of the PAHC questionnaire.We used analysis of variance and Chi square analyses to compare study participants and non-participants on sociodemographic and clinical characteristics. Chi square analysis and Student’s t tests were employed to investigate which sociodemographic and clinical variables, if any, were associated significantly with the PAHC questionnaire domains, the HADS, and the DT. Any variable with a p value below 0.10 was entered subsequently into a logistic (for the PAHC domain scores) or a linear regression model . Only those participants with children completed the domain addressing children-related issues. Thus the analyses relating to this domain were performed on the subgroup of participants with children (n = 100).Chi square and Student’s In total, 263 eligible counselees were invited to participate in the study, of whom 139 (53 %) agreed to do so. Reasons for non-participation included logistical or scheduling problems (n = 23), perceived emotional burden (n = 20), lack of interest (n = 13), and not wanting the counseling session to be audiotaped (n = 3) (audiotaping was employed for another part of the study). Thirty-nine counselees provided other reasons, and 26 did not provide a reason. Two additional cases were excluded from the analysis because their clinical data were not available. This resulted in a total of 137 cases for the analysis. No statistically significant differences were observed between study participants and non-participants on any of the available sociodemographic and clinical variables.The sociodemographic characteristics of the sample are reported in Table Approximately 10 % of the participants did not report any problems included in the PAHC questionnaire that were of a sufficient magnitude  to be considered relevant for further discussion. Fifty-four percent of the participants met the threshold for clinical relevance on three or more domains of the PAHC questionnaire Table . The domAll of the PAHC questionnaire domains were correlated significantly with psychological distress as measured by the HADS, when based on a Pearson correlation coefficient. However, when correcting for inter-domain correlations, only the domains ‘family and social issues’ and ‘general emotions’ remained statistically significantly associated with the HADS. All of the partial correlations were low, with the exception of the domain ‘general emotion,’ which has a strong conceptual overlap with distress as assessed by the HADS Table .The domains ‘hereditary predisposition’, ‘practical issues’, and ‘general emotions’ had statistical significant Pearson’s correlations with distress as measured by the DT. These domains remained statistically significant when correcting for inter-domain correlations. However, the magnitude of the  correlations was relatively low Table .p = 0.001), and having a personal history of cancer (p = 0.03) remained statistically significant. However, only 10 % of the variance in distress scores was explained by these three variables.Education level, having had previous contact with a psychosocial worker, and having a personal history of cancer were associated significantly with general distress as measured by the HADS  and ‘family and social issues’ (p = 0.007). Previous contact with a psychosocial worker  was associated significantly with the domain ‘practical issues’ (p = 0.04). No sociodemographic or clinical variables exhibited statistically significant associations with the domains ‘general emotions’, ‘living with cancer’ or ‘child-related issues’. The variance in the PAHC domain scores explained by these regression models ranged from 2 to 14 %  distressed , 12. AltWe would stress the potential importance of asking counselees about their specific psychosocial problems at the time of cancer genetic counseling, prior to undergoing DNA testing and receiving the DNA test results. Studies of the routine use of patient-reported outcome measures in daily clinical oncology practice have demonstrated their value in enhancing communication between patients and their health care providers –25. We hThere are several limitations of the current study that should be noted. First, only 53 % of those invited to participate in the study actually did so. Although we did not observe any statistically significant differences between participants and non-participants on sociodemographic or clinical background variables, we cannot say with certainty that our sample was entirely representative of the larger population of interest. However, while a small minority of the non-participants (20 of 124 = 16 %) indicated that they thought the study would be too emotionally burdensome for them , the majority of non-participants either reported more neutral reasons  or did not provide a reason. Second, the large majority of study participants was female and was being counseled for hereditary breast and ovarian cancer, reflecting the population of counselees attending the family cancer clinical at the Netherlands Cancer Institute. However, our results cannot necessarily be generalized to those with other hereditary syndromes. As we did not have sufficient statistical power to do so, future studies are needed to determine if the prevalence of psychosocial problems varies significantly as a function of hereditary cancer syndrome and of sex. Third, the PAHC questionnaire was administered either prior to or immediately following the genetic counseling session. This could potentially affect the observed prevalence of psychosocial problems and the associations observed between the PAHC questionnaire and the HADS and DT, and between the PAHC questionnaire and various sociodemographic and clinical variables. However, our analyses indicated that the prevalence of psychosocial problems did not vary significantly as a function of the timing of the questionnaire administration. Fourth, the domains of the PAHC questionnaire were correlated. While this could potentially complicate the interpretation of observed correlations between the PAHC and other measures and variables, the use of partial correlations corrected for this.The study also had several important strengths. First, as indicated above, the study sample was representative of the population undergoing genetic counseling in our clinic. Second, we included a range of sociodemographic and clinical variables that have been frequently used to try to identify those at risk for psychosocial problems and psychological distress. Thus we were able to compare directly the relative value of risk profiles based on background variables with a psychosocial screening questionnaire in identifying those with clinically relevant psychosocial problems.In conclusion, although only a minority of individuals who undergo cancer genetic counseling suffer from high levels of psychological distress, the large majority reports a range of psychosocial problems related specifically to cancer genetic counseling. The PAHC questionnaire is a useful tool for identifying relevant psychosocial problems that merit further attention in clinical practice. Use of such a tool can contribute significantly to enhancing the quality of communication between genetic counselors and their clients, to providing client-centered care, and to addressing relevant psychosocial problems in a timely manner.


I2 = 66.0%) for saturated fatty acids, 0.98  for monounsaturated fatty acids, and 1.00  for polyunsaturated fatty acids intake. Non-significant results were observed in the majority of subgroup analyses stratified by study characteristics and adjustment for potential confounders in analyses of aforementioned associations. In conclusion, results from this dose-response meta-analysis provided limited evidence that dietary saturated, monounsaturated, and polyunsaturated fatty acids consumption was associated with endometrial cancer risk. Further studies, especial prospective designed or pooled studies are warranted to confirm our findings.Epidemiological studies have provided controversial evidence of the association between dietary fatty acids intake and endometrial cancer risk. The continuous update project of World Cancer Research Fund failed to focus on this issue. To address this inconsistency, we conducted this dose-response meta-analysis based on epidemiological studies published up to the end of June 2015 identified from PubMed, EMBASE and Web of Science. Two authors independently performed the eligibility evaluation and data extraction. Random-effects models were used to estimate summary relative risks (RRs) and 95% confidence intervals (CIs). Fourteen epidemiological studies (4 cohort and 10 case-control studies) were included in this dose-response meta-analysis. The summary RR for an intake increment of 10g/day was 1.02 (95% CI = 0.97–1.08;  About 320,000 new cases of endometrial cancer (EC) were diagnosed and nearly 76,000 deaths from this disease occurred worldwide in 2012 [In vitro and in vivo studies have indicated that dietary fat and fatty acids (FA) intake have been proposed to influence EC risk by modulating the production, metabolism, and excretion of endogenous hormones [I2 = 52.7%, P for heterogeneity = 0.06) for the highest compared with the lowest intakes of dietary saturated FA [hormones –7. Howevhormones . The relhormones . In 2007rated FA . Howeverrated FA . During rated FA . In contrated FA . AdditioFigure Table The information of study quality assessment is demonstrated in Table I2 = 66.0%, P for heterogeneity < 0.01) . There was no indication of publication bias by visual inspection of the funnel plot  and Begg's test (P for bias = 0.30).Thirteen studies , 9–20 wet Figure  as well I2 = 15.3%, P for heterogeneity = 0.31). However, the non-significant results with high heterogeneity were observed in the majority of other subgroup analyses  after omitting the study by Littman et al [I2 = 61.4%, P for heterogeneity < 0.01) after omitting the study by Jain et al [When stratified by study design, we found borderline significant result in cohort study , with low heterogeneity  . There was no indication of publication bias by visual inspection of the funnel plot  and Begg's test (P for bias = 0.59).Nine studies , 17–19 w) Figure . No evidt Figure  as well I2 = 0%, P for heterogeneity = 0.64) after omitting the study by Lucenteforte et al [I2 = 0%, P for heterogeneity = 0.81) after omitting the study by Tzonou et al [The direction of the associations was consistent in the subgroup analyses stratified by study characteristics and adjustment for potential confounders. Furthermore, the majority of results showed borderline significance without or with little heterogeneity in the subgroup analyses except for these studies without using energy-adjusted model or without adjustment for oral contraceptive use Table . The reste et al  to a higou et al .I2 = 0%, P for heterogeneity = 0.46) . There was no indication of publication bias by visual inspection of the funnel plot  and Begg's test (P for bias = 0.71).Eight studies , 17, 19 I2 = 0%, P for heterogeneity = 0.47) after omitting the study by Cui et al [I2 = 0%, P for heterogeneity = 0.56) after omitting the study by Merritt et al [When stratified by study characteristics and adjustment for potential confounders, the non-significant results were observed in all the subgroups Table . The restt et al .In this dose-response meta-analysis of 14 epidemiological studies, we found no statistically significant association between intake of saturated FA and EC risk. However, to the best of our knowledge, this is the first meta-analysis to report the relationship between monounsaturated and polyunsaturated FA intake and EC risk.I2 = 52.7%) intake was associated with an increased risk of EC [The present study is inconsistent with a previous meta-analysis which suggested that dietary saturated FA  [n = 3) [vs. 1.06), which could be partly attributed to the methodological differences in study designs. Compared with case-control studies, prospective studies are less susceptible to bias  due to their nature. Additionally, on the basis of the updated NOS, less case-control studies fulfilled these criteria than cohort studies   and phys [n = 3) , 13, 14. [n = 3) ; althoug [n = 3) . In cont [n = 3)  measure In summary, our dose-response meta-analysis provides limited evidence that dietary intake of saturated, monounsaturated, and polyunsaturated FA was associated with the risk of EC. Since few prospective studies were included, the findings of this study are warranted to be confirmed. Additionally, further studies are needed to provide more detail results, including those for other type of dietary FA and stratify the results by the histology of EC after better adjustment for the potential confounding.Two independent investigators (Q-JW and T-TG) systematically searched PubMed, EMBASE, and Web of Science from each database's inception to the end of June, 2015 to identify relevant epidemiological studies. The following search keywords were used: (diet OR dietary OR fat OR fatty OR fatty acid) AND  AND (cancer OR tumor OR carcinoma OR neoplasm). This search strategy was similar to previous studies –25. We fTo be included in this analysis, a study must have (i) an observational study design; (ii) evaluated the association between dietary FA  intake and EC risk; and (iii) presented RR, odds ratio (OR), or hazard ratio (HR) estimates with 95% CIs or necessary data for calculation . If seveThe studies were excluded by the following exclusion criteria: (i) were randomized controlled trials, reviews without original data, ecological studies, editorials, and case reports; (ii) reported the risk estimates that could not be summarized (such as reported the risk estimates without 95% CIs); and (iii) reported the outcome as EC mortality or recurrence .Data were extracted by two investigators (Q-JW and T-TG) using a data extraction form and entered into a database. All differences were resolved by discussion with third investigator (Y-ZW). For each included study, we extracted the following information: last name of the first author, publication year, geographic location, number of cases/controls (size of cohort), exposure assessment and categories, and study-specific adjusted estimates with their 95% CIs for the highest compared with the lowest category of intake (including adjusted confounders information if applicable). If there were multiple estimates for the association, we used the estimate adjusted for the most appropriate confounding variables, like previous studies , 27–29.An update Newcastle-Ottawa Scale (NOS) , 29–31 uAs the absolute risk of EC is low and therefore we interpreted all risk estimates as relative risk (RR) for simplicity . For stuP-value by testing that these two regression coefficients were simultaneously equal to zero. We calculated a P-value for nonlinearity by testing that the coefficient of the second spline was equal to zero. The details of this method has been published elsewhere [To examine the associations between the dietary FA intake and EC risk, the summary RR with 95% CIs were estimated by summarizing the risk estimates of each study using the random effect models, which considered both within- and between-study variation . We summlsewhere , 38.For conducting the dose-response meta-analysis, the following information were needed: (i) the distribution of cases and non-cases and the risk estimates with the variance estimates for at least three quantitative exposure categories; (2) the median or mean level of these exposures in each category . Given this, thirteen, nine, and eight studies met the criteria and were included in the dose-response analysis of saturated, monounsaturated, and polyunsaturated FA intake and EC risk, respectively.versus case-control studies), quality scores (high versus low), geographic location (North America versus Europe), validated food frequency questionnaire (yes versus no), mean number of EC cases (≥450 versus < 450), energy-adjusted model (yes versus no), and adjustment for potential confounders including total energy intake, body mass index, cigarette smoking, parity, oral contraceptive use, menopausal status, and hormone replacement therapy use. Heterogeneity between subgroups was evaluated by meta-regression [To investigate the possible sources of heterogeneity of main results, we carried out stratified analyses by the following study features: study design .Small study bias, such as publication bias can reflect genuine heterogeneity, chance, or other reasons for differences between small and large studies which was evaluated with Egger's linear regression asymmetry test  and Begg


Inflammation is a response of body tissues to injury or irritation. Small RNAs, such as miR-146a and miR-499, participate in various processes of tumorigenesis. A recent study indicates that inflammation and abnormal immune responses may promote malignant progression in cancer development, indicating that inflammation-related polymorphisms such as miR-146a rs2910164 and miR-499 rs3746444 are crucial. However, studies on the association of these two polymorphisms with hepatocellular carcinoma (HCC) are inconclusive and inconsistent. We aimed at accessing the combined result of reported studies and make a more precise estimate of the relationship.Meta-analysis was performed on the associations between the miR-146a rs2910164 C > G and miR-499 rs3746444 T > C polymorphisms and hepatocellular carcinoma, using: allele contrast, dominant, and recessive models. A total of 12 studies(8 on miR-146a rs2910164 and 4 on miR-499 rs3746444) with three populations  were included in this study.P < 0.001; GC versus CC: OR = 1.165, 95% CI 1.054–1.286, P = 0.003; GG versus CC: OR = 1.361, 95% CI 1.192–1.553, P < 0.001; GG/GC versus CC: OR = 1.213, 95% CI 1.104–1.333, P < 0.001; GG versus GC/CC: OR = 1.210, 95% CI 1.080–1.356, P < 0.001). Our data suggest that people with G allele have a higher susceptibility to HCC as compared to those with C allele. However, meta-analysis failed to detect associations between miR-499 rs3746444 and HCC risk under each genetic model tested. Subgroup analysis showed that Chinese population with CC genotype are more vulnerable to HCC  than those with TT genotype.Results show that both allele frequency and genotype distributions of miR-146a rs2910164 polymorphism are significantly associated with susceptibility to HCC (G versus C: OR = 1.153, 95% CI 1.083–1.228, We conclude that rs2910164 in miR-146a may confer susceptibility to HCC, especially in the Chinese population. No significant association was found between miR-499 rs3746444 and HCC, but subgroup study showed that subjects with CC genotype are more vulnerable to HCC than TT genotype in the Chinese population. Hepatocellular carcinoma (HCC) is one of the most common malignant tumors and the third most common cause of cancer-related mortality worldwide , especiaSince microRNAs (miRNAs) were first described in 1993, these small, evolutionarily conserved, endogenous, single-stranded, non-coding RNA molecules , typicalet al. concluded that inflammation in the background non-cancerous cirrhotic portion of the liver could evoke malignant progression in HCC development . He. Heet alWe carried out a computer-based search of the most commonly used database, including PubMed, EMBASE, Web of Science, Cochrane Central Register of Controlled Trials, China National Knowledge Infrastructure (CNKI),China Biological Medicine Database (CBMD) and VIP(Chinese) to identify all studies examining the relationship between miR-146a rs2910164, miR-499 rs3746444and HCC. The following keywords and subject terms were used for searching: “miR-146a”, “microRNA”, “rs2910164”, “miR-499”, “rs3746444”, “polymorphism”, “Hepatocellular Carcinoma”, “liver cancer”, “HCC”,” Liver neoplasm” as well as combinations, without any restriction on publication date and language by two independent investigators (Jiajing Liu and Bo Xie) up to February, 2013. References of retrieved articles were also scanned. Corresponding authors of published papers without sufficient data were contacted by e-mail for more information. The design of the meta-analysis meets the requiries of moose checklist  the study followed a case–control design and the participants were in both groups voluntarily joined this study with informed consents; (ii) the study was designed to evaluate the associations between the SNP miR-146a rs2910164 or miR-499 rs3746444 polymorphisms and susceptibility to HCC; (iii) participants recruited from comparable populations with similar demographic background in control and case groups; (iv) a proper diagnosis criteria of HCC to determine the disease status respectively; (V) detailed genotype data were provided for the calculation of odds ratio (OR) and 95% confidence interval (95%CI); (vi) proper method to determine the genotypes with detailed description of method used; (vii) Hardy-Weinberg equilibrium (HWE) reached.We excluded the following: 1) studies that contained overlapping data, the latest study was included; (2) studies in which the number of wild genotypes could be ascertained; (3) the manuscript was published as a review or abstract, not a full original paper. The details are demonstrated in Figure  studies P = 0.05, P < 0.05 was considered a departure from HWE). If the control groups were not in HWE, sensitivity analysis or subgroup analysis was performed to test the robustness of the findings.The Hard-Weinberg Equilibrium (HWE) was evaluated by the goodness-of-fit chi-square test for the control groups of each study  were two-sided.The odds ratio (OR) corresponding to the 95% confidence interval (95% CI) was calculated to assess the strength of association between rs2910164 and rs3746444 polymorphisms and susceptibility to HCC. We performed a meta-analyses using allele frequency comparison; additive model; dominant model; and recessive model. The significance of pooled ORs was determined by  studies , so the  studies  was appl studies  are moreion bias . All anaet al. ,,. In the P < 0.001; GC versus CC: OR = 1.165, 95% CI 1.054–1.286, P = 0.003; GG versus CC: OR = 1.361, 95% CI 1.192–1.553, P < 0.001; GG/GC versus CC: OR = 1.213, 95% CI 1.104–1.333, P < 0.001; GG versus GC/CC: OR = 1.210, 95% CI 1.080–1.356, P < 0.001). The forest plots under additive model of study between the miR-146a rs2910164 polymorphisms and HCC risk are presented in Figure The results of meta-analysis of the miR-146a rs2910164 polymorphism and HCC risk are shown in Table P = 0.577) and PCR-RFLP method  subgroups. Though not statistically significant, the PCR-RFLP method still identified G allele as a risk factor (OR > 1) in comparison with C allele. However, other genotyping methods , such as MassArray, still found an association between SNP and HCC.In addition, a subgroup analysis stratified by study characteristics like population (Chinese and other), genotyping method (PCR-RFLP and other), sample size, and significant association was performed for a better understanding of the relationship between the SNP and HCC Table . The stuP = 0.554; CT versus TT: OR = 0.994, 95% CI 0.784–1.262, P = 0.963; CC versus TT: OR = 1.266, 95% CI 0.596–2.687, P = 0.540; CC/CT versus TT: OR = 1.091 95% CI 0.726–1.641, P = 0.674; CC versus CT/TT: OR = 1.265 95% CI 0.924–1.732, P = 0.143).The results of meta-analysis based on 4 independent population samples with 746 patients and 1006 controls for the miR-499 rs3746444 polymorphism and HCC risk are shown in Table P = 0.017) in the Chinese population. Subgroup analysis did not show significant heterogeneity under allele frequency contrast model, except in the Chinese subgroup. Although meta-regression analysis failed to identify the source of homogeneity due to the limited number of studies, the subgroup analysis indicated population and sample size as the two major concerns for the source of heterogeneity.Furthermore, we conducted subgroup analysis including those on population (Chinese and other) and sample size (<300 and >300), to analyze characteristic homogeneous groups in all the genetic models and assess heterogeneity among the studies Table . The subP values for bias of >0.05). The result confirmed the absence of publication bias in this meta-analysis  variation in miRNA genes has been considered uncommon owing to high evolutionary conservation across species . These set al. . Al. AlP = 0 ,,, irrespe in 2010 , two wel ,[43,[43et al [et al.  showing  [et al.  study, sSeveral limitations of this research should be considered when interpreting the results due to some limitations of this meta-analysis. Firstly, our literature searching was depended on English and Chinese databases only. Language bias may exist. Secondly, our study only included three populations , so the result may not be able to extrapolation to other populations such as African. Thirdly, lack of available data of many environmental factors including age, HBV/HCV infection status, gender and alcohol consumption may limited the possibility of a more comprehensive evaluation of the association between the SNP and susceptibility to HCC. Fourthly, no significant association was found between miR-499 rs3746444 and HCC under every models may due to the relatively small sample size of those four studies, and should be interpreted with caution. Lastly, even to our best knowledge, we can’t discover all grey literatures.We performed comprehensive literature search in multiple databases without limiting publication language and date. We conducted quantitative data synthesis in allele frequency, additive model, dominant model and recessive model. Subgroup analysis was also performed by different characteristics among studies like region distribution, genotyping method and sample size under the additive models. Q-test in all subgroups showed no significant heterogeneity. Our meta-analysis, though with limitations, concludes that there is a strong significant association between rs2910164 in miR-146a and susceptibility to HCC, especially in Chinese population. No significant association was found between miR-499 rs3746444 and HCC under every models, but subgroup study shows subjects with CC genotype in Chinese group are more vulnerable to HCC  than those with TT genotype. This result should be interpreted with caution. Well-designed studies with more ethnic groups are required to further validate the results.The authors declare that they have no competing interests that are directly relevant to the content of this study.JL was in charge of conceived and designed the study. JL and BX were responsible for collection of data and performing the statistical analysis and manuscript preparation. FJ and SC were responsible for checking the data. WM participated in study design and critically revised the manuscript. All authors were responsible for drafting the manuscript, read and approved the final version.Moose checklist.Click here for file


Toxicity caused by chemical mixtures has emerged as a significant challenge for toxicologists and risk assessors. Information on individual chemicals' modes of action is an important part of the hazard identification step. In this study, an automatic text mining-based tool was employed as a method to identify the carcinogenic modes of action of pesticides frequently found in fruit on the Swedish market. The current available scientific literature on the 26 most common pesticides found in apples and oranges was evaluated. The literature was classified according to a taxonomy that specifies the main type of scientific evidence used for determining carcinogenic properties of chemicals. The publication profiles of many pesticides were similar, containing evidence for both genotoxic and non-genotoxic modes of action, including effects such as oxidative stress, chromosomal changes and cell proliferation. We also found that 18 of the 26 pesticides studied here had previously caused tumors in at least one animal species, findings which support the mode of action data. This study shows how a text-mining tool could be used to identify carcinogenic modes of action for a group of chemicals in large quantities of text. This strategy could support the risk assessment process of chemical mixtures. Chemical risk assessment of mixtures is an important but challenging task for toxicologists. Unlimited variations of mixtures in our environment and knowledge gaps about toxic effects caused by chemical mixtures are examples of factors that make this process complex. Mixture effects can be described as caused either by additivity or interactions (such as synergistic or antagonistic effects) of the individual compounds  reports that the percentage of samples of fruits, vegetables and cereals with multiple residues increased by 11 percent from 1997 to 2007 (from 15 to 26 percent). In 2008, residues of two or more pesticides were found in 27 percent of the samples analyzed. The same proportion, 27 percent of samples containing multiple residues, was found in 2010. One sample of grapes was found to contain as many as 26 different pesticide residues   was gathered via a search (in August 2013) using the names of the pesticides. A computational tool, CRAB, was used to analyze the abstracts. The tool classifies PubMed abstracts automatically according to a MOA taxonomy. The MOA taxonomy captures the current understanding of processes leading to carcinogenesis and is based on two main categories: genotoxic and non-genotoxic MOA. The taxonomy is further structured into sub-categories according to a classification of Hattis et al.  are listed in Table We investigated the literature of human and animal Data from an analysis of apple and orange samples carried out in 2009–2010 was kindly provided by the Swedish NFA. Data shows that the majority of the tested apples and oranges contained several pesticide residues, 78 percent of the apple samples contained more than one pesticide and for orange samples this number was as high as 96 percent. The results thus show that two or more pesticide residues were freqently detected in apple and orange samples. For example, one apple sample contained residues of seven pesticides and in one orange sample 10 different pesticide residues were detected.Table The distribution of abstracts over the MOA taxonomy was analyzed in detail. The distribution of classified abstracts for the individual pesticides in apples and oranges is shown for 11 selected common MOA categories Figures . The disWe compared the MOA analysis by the CRAB tool with carcinogenic evidence and classifications for each pesticide Table . Common When all information on cancer classifications and published studies of the 26 pesticides was summed, it showed that the majority of the chemicals have evidence or classifications that suggest carcinogenic potential. The animal tumor data retrieved from published literature through PubMed shows evidence of carcinogenicity reported previously for 18 of 26 substances  for all pesticides permitted in food products intended for human consumption EFSA, . IndividThe CRAB tool has many advantages over manual literature analysis when large quantities of data need to be examined. The tool provides a rapid view of published literature and can point to carcinogenic MOA that groups of chemicals can have in common. We have previously conducted case studies to demonstrate how the text-mining tool can be used to support cancer risk assessment. For example, literature profiles of well-known carcinogens were compared with the known properties of each chemical and the classification results correlated well to what was previously known about these substances (Korhonen et al., Hazard identification and risk assessment of mixtures is a complex and challenging process. The study described here provides an example of how a text mining-based tool could support the analysis of large amounts of textual information to detect trends and patterns in data. A MOA analysis can identify common links between different chemicals which could serve as basis for hypothesis generation and direct further research and risk assessment of chemical mixtures.All authors were involved in the design of the study. Ilona Silins conducted the text mining experiments, analysis and the manual literature review. All authors contributed to the writing of the manuscript. All authors approved the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


In the last decade, ample evidence has demonstrated the growing importance of androgen receptor (AR) CAG repeat polymorphism in andrology. This genetic parameter is able to condition the peripheral effects of testosterone and therefore to influence male sexual function and fertility, cardiovascular risk, body composition, bone metabolism, the risk of prostate and testicular cancer, the psychiatric status, and the onset of neurodegenerative disorders. In this review, we extensively discuss the literature data and identify a role for AR CAG repeat polymorphism in conditioning the systemic testosterone effects. In particular, our main purpose was to provide an updated text able to shed light on the many and often contradictory findings reporting an influence of CAG repeat polymorphism on the targets of testosterone action. The main mechanism of action for AR is direct regulation of gene transcription. After the binding of an androgen to its receptor, a conformational change occurs, causing the dissociation of heat shock proteins, translocations into the nucleus, and dimerization . The AR eductase .The AR gene is composed of eight exons and is located on X chromosome at q11-q12. Exon 1 of the AR gene contains a polymorphic sequence of CAG repeat, which usually varies in number from 10 to 35, and which encodes polyglutamine stretches of AR transactivation domain . In factEthnic differences in CAG repeat length of AR gene exist between African, Caucasian, and Asian populations. Evaluating ethnic groups is important to understand the variability of AR gene and the differences in androgen sensitivity in the general population . The allIn this review, we aim at discussing the literature data indicating a role for AR CAG repeat polymorphism in conditioning the peripheral effects of testosterone. In particular, our main purpose was to provide an updated contribution able to shed light on the many and often contradictory findings on the influence of CAG repeat polymorphism on the targets of testosterone action .The relationship between CAG repeat polymorphism and sexual function has not yet been well explored. In fact, few studies have examined this issue and knowledge on this aspect is incomplete. As far as transversal studies are concerned, a discrepancy on the role of CAG polymorphism emerges, probably due to the lack of homogeneity in recruited samples and methodological procedures. Pastuszak et al. reviewed the medical records of 85 men who presented to their clinic. AR gene CAG repeat number was found to be negatively correlated with all domains of sexual function assessable by IIEF-15 . ConversRegarding longitudinal studies, only two reports evaluated the effects of the polymorphism in conditioning sexual function recovery after testosterone replacement therapy in hypogonadal male patients. They both found that shorter CAG length is associated with a greater improvement of several aspects of IIEF questionnaire after TRT , 13. HowDiagnostic andrology has seen considerable progress in recent years. It is now possible to establish the cause of a significant percentage of cases of infertility previously considered idiopathic. For example, it has been demonstrated that genetic factors are responsible for 15% of male infertility . The corNumerous studies in recent years have attempted to establish the relationship between CAG repeat length variation and male infertility to find out if this variability in the AR gene could be associated with impaired spermatogenesis. Although it is now known that androgens and the AR receptor gene contribute toward correct sexual differentiation and maintenance of normal spermatogenesis, discordant results are found in the literature. In Australian, North American, and Japanese populations, an association between CAG repeat length and male infertility has been reported –18, but In 2004, Ferlin et al. studied 163 infertile and 115 normozoospermic fertile men and found no significant difference in the mean number of CAG repeats between the two groups. The authors reached the same result when subdividing the 163 infertile patients by semen phenotype . The authors found a statistically significant difference between cases and controls when analysing the combination of CAG with GGC polymorphism: infertile patients most commonly demonstrated the combination CAG = 21/GGC = 18, and very rarely CAG ≥ 23/GGC ≤ 16. Furthermore, patients with severe oligozoospermia most frequently demonstrated the combination CAG ≥ 21/GGC ≥ 18. The authors concluded that the number of CAG and GGC triplets has a combined effect on the AR receptor function and that this was the first evidence of a relationship between particular CAG/GGC haplotypes and male infertility [In 2010, Castro-Nallar et al. found no statistically significant difference in the mean number of CAG repeats between 117 infertile patients (with azoospermia and oligozoospermia), of whom 93 were with idiopathic infertility and 121 normozoospermic controls. The authors suggest that the CAG triplet allele 21 might increase the risk of Sertoli cell-only syndrome, although the mechanism is not yet clear [Nenonen et al., in a 2010 meta-analysis of 3915 men (1831 fertile and 2084 infertile), demonstrated that there was no statistically significant difference in the mean number of CAG repeats between the two groups. However, on dividing the CAG repeat length into three groups , a binary logistic regression analysis found that men with <22 and >23 CAG repeats had an approximately 20% higher risk of being infertile than men with 22 or 23 CAG repeats [In 2013, Han et al. conducted a cytogenetic and molecular study on a population of infertile Chinese men. The authors analysed CAG polymorphism in 101 azoospermic and 54 oligozoospermic patients and 96 controls, excluding patients with Y chromosome deletions and chromosome abnormalities from the study. The authors did not find any statistically significant difference in CAG numbers between cases and controls. When analysing the joint association CAG > 23 and GGC < 23, the authors found a statistically significant difference between cases and controls only in GGC repeats .It is known that the AR gene CAG repeat stretch is found in the transactivation domain of the protein that interacts with the domain by binding androgens. The fact that mutations in the CAG repeat stretch can block this interaction , 26 or cThe prostate is an androgen-regulated organ and androgens also play a key role in the pathogenesis of prostate cancer. In animal models the use of androgens is essential to induce prostate cancer and men and dogs castrated before puberty do not develop prostate cancer. In addition the reduction in testosterone levels is therapeutically used to slow the growth of prostate cancer .It is still unclear how androgens are involved in the etiology of prostate cancer. They are able to stimulate the growth of an existing malignant process, but there is no evidence that they might trigger the cancerous process by facilitating the transformation of a benign cell into a malignant one .It is likely that the genesis of prostate cancer is not induced by androgens, but that the stronger androgenic stimulation caused by receptors with shorter CAG repeats may contribute to a faster development of malignant cells. This could explain the higher risk of cancer and the earlier age of onset .An experimental study showed that in prostatic epithelial cells an inverse relationship exists between AR transactivity and polyglutamine tract length . Some stExperimental data also suggest that the increased transcriptional activity favors the formation of TMPRSS2:ERG, a fusion gene found in about 50% of prostate cancers. In the same study, which evaluated 291 men with prostate cancer (147 ERG-positive) and 1.221 cancer-free controls, subjects with shorter CAG repeats had an increased risk of ERG-positive (odds ratio (OR), 1.07 per 1 repeat decrease; 95% CI, 1.00–1.14), but not ERG-negative prostate cancer . These data suggest that short CAG repeats are associated with the development of TMPRSS2:ERG-positive prostate cancer . On the Many authors have suggested that the CAG repeat length is inversely correlated with the risk of developing prostate cancer . CoetzeeHowever, it must be acknowledged that not all studies agree in concluding that shorter CAG repeats are associated with an increased risk of cancer. While some works fail to show a statistically significant association between shorter CAG repeats and prostate cancer , 40, othP < 0.02) trend for large adenoma size with short CAG repeat length among the adenoma quartiles, thus demonstrating the inverse relationship between prostatic adenoma size and AR gene CAG repeat length [Finally, shorter CAG repeats seem also to be associated with the development of benign prostatic hypertrophy (BPH). According to data from the Health Professionals Follow-Up Study, men with AR gene CAG repeat lengths of 19 or less had an OR of benign prostatic hyperplasia of 1.92 relative to men with repeat lengths of 25 or more . Also, at length .The role of testosterone in the cardiovascular health of men is controversial. Some evidence exists showing that hypogonadism could be associated with a worse metabolic profile and cardiovascular risk ; howeverA recent study on 1859 men aged 20–79 years showed no direct correlation between CAG repeat length of AR and cardiometabolic risk factors . Other aOn the other hand, independent associations between CAG length and adverse cardiovascular risk factors, such as high LDL , low HDLA possible relationship has been suggested between body composition and the CAG repeat polymorphism of the AR gene. In adolescent boys, low CAG repeat numbers in AR may be a genetic risk factor for fat accumulation , particuThe influence of CAG repeat polymorphism in the AR gene on bone health is also not clear. CAG length of the AR gene was found to have a negligible –64 or poTesticular cancer (TC) accounts for 1% of all cancers in men. It comes in a broad variety of histotypes: over 90% of cases originate from the germinal epithelium of the seminiferous tubules, making this the largest group of testicular cancers. Its incidence varies from 1 per 100,000 in Asia and Africa to 9.2 per 100,000 in Denmark –73. The Irvine suggested that a longer CAG repeat region could reduce the receptor's transactivation activity . Many auP < 0.05) [Garolla et al. analysed 123 stage 1 testicular cancer patients against a control group of 300 fertile men studied for AR mutations, of whom 115 were selected for the CAG and GGC repeats study . There w < 0.05) .In 2012, Kristiansen et al. investigated the correlation between CAG repeat length and testicular cancer in a Norwegian population of 651 TC patients and 313 controls. No statistically significant differences were seen in the number of CAG repeats between patients and controls, even when analysed by histotype . In addiPrevious studies have correlated CAG repeat length with clinical stage of testicular cancer, reporting higher CAG repeat numbers if the tumour was advanced at diagnosis . GrassetIn conclusion, a CAG repeat number of ≥25 may be considered a risk factor for the onset of testicular cancer, given its greater frequency in patients compared to controls . This isThe role of sex hormones in the genesis of TC has also roused considerable interest in recent years. Various studies have investigated exposure to endocrine disrupter chemicals, which may be associated with testicular cancer. Interest has focused on the synergism between gene modifications and the influence of the environment as a possible risk factor for the onset of testicular cancer. It was recently suggested that postnatal exposure might also increase the risk of developing TC and that androgen secretion during puberty might be involved in TC progression .In 2008, a case-control study  investigHowever, comparative studies of groups of single ethnic origins are still lacking, since there is a highly variable distribution of these polymorphisms in different populations worldwide. This would enable further understanding of the role of the AR gene and polymorphism frequency in the onset of testicular cancer in patients of different ethnic origins.Important differences in behavior, personality, and depressive tendencies exist between males and females. Men seem more inclined to search dominance  and to tSome authors have shown an association between shorter CAG repeats and self-report measures of dominance and prestige, all of which are argued to be indices of the mating effort . Others Regarding depression, discordant findings exist. Some studies show an association between longer CAG repeats and higher levels of depression , while oIn addition, violent behavior  and impuCAG repeat may be involved in the pathogenesis of some neurodegenerative diseases . HoweverIn 2015, a large body of evidence indicated an important role for AR CAG polymorphism in conditioning the peripheral effect of testosterone, even if its contribution warrants further assessment because of the many controversial findings in each androgen-related action. Of note, other associations are emerging , but thAt present, AR CAG polymorphism is not recommended in the routine setting. However, in the near future, it could become of clinical relevance because of the theoretical possibility of identifying subjects more or less at risk for various disorders and more or less responsive to testosterone treatment. In this last case, study of CAG repeat length could allow us to individually tailor testosterone replacement therapy, as subjects with shorter CAG repeat could need lower doses of testosterone while men with longer repeats could require higher ones.


CD143) gene was reported to repress cough reflex by reducing bradykinin and substance P levels, thus increasing the likelihood to develop pneumonia. There have been different reports regarding the correlation of CD143 rs4340 genotypes with pneumonia risk, which prompted us to perform a meta-analysis to determine the elusive association.rs4340 polymorphism at intron 16 of the angiotensin-converting enzyme  and 95% confidence intervals (CIs) were calculated to assess the risk of pneumonia. The fixed-effects model (FEM) was used.We combined multiple keywords to identify the studies addressing the association between A total of 10 studies were analyzed in this quantitative analysis. We found a strong association between rs4340 single nucleotide polymorphism (SNP) and pneumonia risk using the recessive model . A significantly increased risk was also indicated under the recessive model in Asian populations , community-acquired pneumonia (CAP) , and nosocomial pneumonia (NP) .CD143 rs4340 polymorphism may represent a risk factor for pneumonia.Our meta-analysis demonstrates that  Angiotensin-converting enzyme (CD143) is such a regulatory enzyme essential for the conversion of angiotensin-I (AT-I) to AT-II, a product in the process of proteolytic cleavage of angiotensinogen.7CD143 inhibitors have been shown to decrease the risk to develop pneumonia in elderly individuals and a reduction in mortality rate has been reported in patients who use CD143 inhibitors.9 These data indicated that a clearer understanding of the correlation between CD143 and pneumonia will help prevent or survive this disease.Available data implicate that pathogenesis of pneumonia correlates with attenuation of cough reflex,CD143 gene on chromosome 17q23 spans over 24 kb in length and contains 26 exons. Serum CD143 levels are genetically determined and around half of the variance arises as a result of a functionally important insertion/deletion  polymorphism (rs4340) at intron 16. An association of the D-allele with enhanced CD143 activity has previously been established.10 The same allele has been linked with a wide range of diseases, ranging from cardiovascular events and renal disorders.11 Moreover, individuals harboring the D-allele are shown to have greater risk of pneumonia.12 The I/I genotype, however, contributes to a high chance of survival for patients with acute respiratory distress syndrome.14The human 16 We therefore performed a meta-analysis, attempting to clarify whether or not the rs4340 SNP is correlated with risk of pneumonia by means of meta-analysis.Published epidemiological studies reported different conclusions regarding the role of rs4340 single nucleotide polymorphism (SNP) in pneumonia predisposition.This study was approved by the PLA 307th Hospital Ethics Committee. This study does not involve patients, so ethical approval was not required.ACE,” “CD143,” “polymorphism,” “genotypes” “variants,” “pneumonia,” “community-acquired pneumonia,” and “acute respiratory distress syndrome” to identify the case-control studies covered in the EMBASE, Google Scholar, PubMed, and CNKI databases. All research articles addressing the correlation of CD143 rs4340 SNP with pneumonia risk were checked for their references to obtain additional data. We also manually searched the journals known to publish pneumonia-related articles. The last search was undertaken in October 2013.We combined “angiotensin-converting enzyme,” “CD143 rs4340 SNP with pneumonia risk was evaluated; genotype distribution must be clearly presented; the case group consisted of at least 30 patients. Any study that violated the aforementioned requirements was certainly excluded from this analysis. When more than 1 study employed the same cases, we selected the study with the largest number of subjects.Eligible studies were selected based on: pneumonia patients were investigated; reference group consisted of healthy individuals; the association of Two investigators strictly followed the inclusion criteria to select all eligible studies. Data extraction was independently performed. Information on authors’ surname name and institute, publication year, study country, racial descent (ethnicity), source of controls, determination of rs4340 polymorphism, type of pneumonia, number of cases and controls, and genotype frequency was collected for each study. Racial descent was grouped as Caucasian, Asian, or mixed when ethnic origin was unclear; type of pneumonia was classified as community-acquired pneumonia (CAP), nosocomial pneumonia (NP), or information inaccessible (II) when data were not stated. Disagreements were resolved through discussion among all investigators and a final decision was based on major opinions.Both overall and stratified meta-analyses (by ethnicity and type of pneumonia) were performed in this study. Using the allele and genotype data, we assessed the risk of pneumonia  with an assumption of D/D versus I/I (homozygote model), I/D versus I/I (heterozygote model), D/D + I/D versus I/I (dominant model) and D/D versus I/D + I/I (recessive model).17 and P values less than 0.05 were considered significant. In such a case, the random-effects model (REM) derived from the DerSimonian–Laird method was used to summarize the values for each study,18 or else the fixed-effects model  was applied.19 Begg funnel plots and Egger linear regression test were utilized to determine publication bias.21 Sensitivity analysis was performed to check if the single studies had significant influence on the overall meta-analysis results. The genotype distribution in control groups was checked for Hardy–Weinberg equilibrium (HWE) deviation.Between-study heterogeneity was evaluated by Cochran Q statistic,P values were considered significant when P < 0.05.All statistical analyses were done using the STATA version 12.0 . Two-tailed CD143 inhibitors or serum CD143 levels and pneumonia risk, or subjects unrelated to the polymorphism of interest. After reading the full texts of the remaining articles, we further excluded 2 studies, of which 1 was a case-only study,22 and the other was a duplicate of 2 studies analyzed in the present analysis.23 Therefore, 10 original articles concerning the correlation of rs4340 SNP with pneumonia were considered eligible for this analysis.31 Main characteristics, including first author, publication year, study country, ethnicity of the populations included, assays used in determining rs4340 SNP, source of controls, specific type of pneumonia, and sample size are presented in Table P < 0.001).As depicted in Figure We performed a meta-analysis among 1366 cases and 2575 controls in this quantitative analysis, as shown in Table In the stratified analysis according to ethnicity, we found a significantly increased pneumonia risk under the recessive model in Asian populations . Stratification analysis by type of pneumonia showed that rs4340 was associated with notably increased risk of both CAP  and NP .The independent studies, including the Chinese study with HWE exclusion, were sequentially excluded to examine if the recalculated ORs, compared to the primary combined effects, were significantly different. We found no notable changes throughout the analysis, suggesting our results are stable (figure not shown).P = 0.721; Egger, P = 0.975, the recessive model)  Figure .CD143 gene with pneumonia risk in the past decade. The previous conflicting reports, however, fail to provide compelling evidence, thereby making the association between rs4340 SNP and pneumonia risk elusive. For instance, a prospective case-control study demonstrated data that did not support a causal association of rs4340 genotypes with pediatric pneumonia.16 A most recent study in a Russian population suggested that rs4340 genotypes accounted for the increase in vulnerability to CAP and NP.15 The discrepancy may result from different study designs and distinct ethnic groups. Other possible factors such as the number of subjects analyzed and differences in clinical features and methodology may also cause the inconsistency in results.Multiple lines of work have consistently correlated rs4340 in the 15 However, both CAP and NP are etiologically heterogeneous and involvement of confounders such as gender, diabetes mellitus, duration of surgery, mechanical ventilation, obesity, and urinary tract catheterization may further complicate the etiology of these diseases.33 Therefore, the exact role of rs4340 genotypes play in pneumonia requires further validation.We performed a meta-analysis in this study, attempting to determine the association between rs4340 polymorphism and pneumonia risk. We analyzed a total of 3941 subjects from 10 case-control studies. When all data were combined, we found rs4340 D/D genotype was associated with 1.33-fold increased risk of pneumonia. An obviously higher risk was observed in Asian samples in stratified analysis by racial descent. We also noted that individuals with the same genotype had notably greater risk of CAP and NP, which is consistent with the study by Salnikova et al. 34 In this work, the authors combined 1431 cases and 3600 controls, of which 977 samples should be excluded (the reason was described in study inclusion and characteristics). In addition, the data included were not all in accord with those provided in original articles. It is the above-mentioned reasons along with the current discrepancy in results that prompted us to perform a meta-analysis. We noted that the association indicated in the previous analysis was relatively stronger compared to this study, suggesting rs4340 genotypes may have substantial effects on pneumonia risk.A recently published meta-analysis concerning the same topic suggested a significant association with pneumonia risk.36 D/D genotype was also reported to increase serum levels of the proinflammatory AT-II.37 It is known that reduced bradykinin and substance P levels repress cough reflex and therefore increase the risk to develop pneumonia. Based on these data, we infer that the D/D genotype of rs4340 polymorphism is associated with increased risk of pneumonia, an inference in line with the positive associations revealed in the present study.The findings in our study are supported by accumulated functional evidence. Brown et al provided evidence that rs4340 genotypes induce cough by suppressing the tissue levels of bradykinin and substance P; the I/I and I/D genotypes had a higher cough reflex relative to the D/D genotype.P < 0.05). Due to the significantly higher allele frequency, Caucasians are theoretically more susceptible to pneumonia than Asians, which is inconsistent with the current findings. Third, deviation from HWE probably caused by methodological errors may affect the results in original article and consequently biases the meta-analysis results.Our study shares the same limitations with most meta-analyses. First, publication bias arises due to many reasons and 1 of them relates to inclusion of small studies. We included several studies with a small number of subjects, leading to possible publication bias that often affects the precision of overall results. Second, there is statistically significant difference in D allele frequency between Caucasians and Asians (53.14% vs 38.42%, CD143 gene and pneumonia vulnerability. A similar trend was indicated in subgroups of Asian populations, CAP and NP. Future studies are urgently needed to address the pathogenic role that remains poorly understood.In conclusion, this meta-analysis shows evidence supporting a causal association between rs4340 SNP in the


TCF7L2 gene polymorphism and DN risk is still unclear. To evaluate the relationship, we performed this meta-analysis.It is assumed that genetic factors may participate in the development of diabetic nephropathy (DN). The association between Eligible relevant studies were searched and selected from PubMed, Embase, and ISI Web of Science. Summary effect estimates were derived using a random effects model, with attention to study quality and publication bias. Ethnical approval was not necessary, because this meta-analysis was based on published articles, and did not involve patient consent.TCF7L2 gene polymorphism and DN risk  = 1.10–1.56, Pheterogeneity < 0.00001, P = 0.002). Subgroup analysis showed similar results in Asian , in Caucasian , in rs7903146 mutation , However, no association was observed in Negroid .A total of 7 studies were identified. Analysis of all studies indicated significant association between TCF7L2 gene polymorphism may contribute to the risk of DN. However, more studies should be launched in the future.Our results suggest that  End-stage renal disease (ESRD) is the most severe form of chronic kidney disease, also known as Stage 5 chronic kidney disease or kidney failure. DN is the leading cause of ESRD in the Europe, Japan, and the United States.3 DN was also the major cause for 40.4% of newly developed ESRD in the United States.5 In the UK, it was estimated that around one third of patients with ESRD were due to diabetes.6 In Korea, according to the Fifth Korea National Health and Nutrition Examination Survey in 2011, the prevalence of DN was 26.7%.7 In China, the Shanghai Diabetic Complications Study reported that the prevalence of DN was 26.2%.8 The expenditure for people with DN are extraordinarily high. In the Medicare population alone, DN-related expenditures among this mostly older group were nearly $25 billion in 2011.9 Therefore, earlier diagnosis and prevention of DN become imminent. However, the mechanism of DN is complicated, in addition to the risk of poor glycemic control and hypertension, inherited factors are considered to play an important role in its progression.The incidence and prevalence of diabetes mellitus (DM) have increased sharply throughout the world. According to the latest report, the prevalence of DM has increased to 11.7% in ChinaTCF7L2) gene is located on chromosome 10q25.3,10 which encodes a transcription factor with a high-mobility box and functions in gene activations related to downstream events of the Wnt signaling pathway. TCF7L2 has been revealed as an susceptibility gene for type 2 diabetes.11 Selective disruption of TCF7L2 in the pancreatic β cell impairs secretory function and lowers β cell mass.12 Clinical studies suggested that TCF7L2 gene impaired glucose tolerance through effects on glucagon as well as insulin secretion.13 Several meta-analysis18 also concluded that TCF7L2 gene polymorphisms were associated with DM.Transcription factor 7-like 2  induced TCF7L2 expression through transforming growth factor-β (TGF-β), and TGF-β could induce TCF7L2 translocation from the cytoplasm to the nucleus. Then, TCF7L2 bounded to the activin receptor-linked kinase1 (ALK1) promoter, resulting in increased ALK1 expression. Increased ALK1 enhanced the effect of TGF-β and further promoted the phosphorylation of Smad1. These interactions caused the phenotypic changes in mesangial cells, thereby resulting in the development of glomerulosclerosis. They concluded that the AGEs/TGF-β/TCF7L2/ALK1/Smad1 signaling pathway plays a key role in the development of DN. In order to draw accurate conclusion, we performed this meta-analysis.However, the relationship between TCF7L2 polymorphism and DN is inconclusive. Based on their studies, Bodhini et alTCF7L2, gene polymorphism, including all alternative locations and combinations of the terms. In addition, hand searching was also performed on the reference lists of all included studies as well as reviews recently published. If there were several papers coming from the same study, only the most recent or complete study was included. Each potentially eligible study was evaluated independently by 2 authors (Fan and Meng).We searched and selected from PubMed, Embase, and ISI Web of Science. Last search was updated on October 2015, the following terms were used: DM, DN, Two authors (Fan and Meng) searched and selected the eligible studies, during the process, the following inclusion criteria were used: firstly, the studies should be designed as case-control or cohort study. Secondly, in the studies, there should be sufficient genotype data to calculate 95%confidence intervals (CI) for risk ratios or Odds ratios (ORs). Thirdly, last but the most important is there should be an appropriate description of TCF7L2 polymorphisms in DN patients and controls in these eligible articles. We excluded some studies, the exclusion criteria include studies on animals; studies without the raw data of genotype of TCF7L2; and case reports, editorials, and review articles.24For each included studies, these data were extracted, including the first author, year of publication, country or area, ethnicity, characteristics, allele frequencies with TCF7L2 genotype in DN patients, and controls. We did not exclude certain studies because of the limited number of cases. This meta-analysis was guided to conduct by the PRISMA statement for preferred reporting of systematic reviews and meta-analysis.TCF7L2 gene polymorphism and DN risk. Allele frequencies were estimated, the ORs were conducted for the allele contrast. Heterogeneity was evaluated using I2 statistic, which takes values from 0% to 100% with higher values indicating greater degree of heterogeneity. The value greater than 50% was considered as significant heterogeneity. If there was significant heterogeneity, the random effects model would be used to analyze the pooled ORs25 or analyzed by the fixed effects model.26 To discern the potential reasons of heterogeneity, subgroup analyses were performed by group studies with similar characteristic, such as ethnicity and the same loci rs7903146.We used ORs with 95% CIs for genotypes and alleles to assess the strength of association between P values were 2-sided.The sensitivity analyses were performed to evaluate the reliability of the results by excluding 1 study each time and rerunning the analysis. An estimate of potential publication bias was carried out by inverted funnel plot, Egger test, and Begg test. All statistical tests were performed by RevMan software . Sensitivity analysis, Begg test, and Egger test were accomplished by Stata . All the 31 were excluded because of insufficient data. Overall, 736 articles were included in our meta-analysis. All the included articles were case-control studies. We calculated the allele frequencies in the DN group and control group. The minor allele frequencies were 3815 in DN group, 4303 in control group, total allele frequencies were 11,698 in DN group, and 16,034 in control group. Of the 7 studies, there were three35 with Asian ethnicity, two33 with Caucasian ethnicity, two36 with Negroid, respectively. Six articles36 probed the association of rs7903146 mutation and DN risk. Main characteristics for all eligible studies were shown in Table The process of study selection for inclusion in this meta-analysis is present in Figure Pheterogeneity < 0.00001, P = 0.002). The estimate of I2 showed significant heterogeneity (I2 = 89%) among the studies, therefore, we used random effects model in this meta-analysis. To discern the potential reasons of heterogeneity, subgroup analyses were performed. Stratified analysis by ethnicities, significant association was observed in Asian , and in Caucasian , but not in Negroid   (Figure Begg funnel plot and Egger test were performed to assess the publication bias. No evidence of publication bias was detected in Begg test  Figure  and Egge2 DN is 1 of the most common microvascular complications of DM, and is the leading cause of ESRD in some countries. The pathogenesis of DN is complicated, besides poor glycemic control and hypertension, genetic susceptibility also play an important role. TCF7L2, also known as transcription factor 4 , has been revealed as a susceptibility gene for type 2 diabetes.11 However, the association between TCF7L2 gene and DN risk is elusive. To derive a more precise estimation of the relationship, we performed this meta-analysis.Recent epidemiological studies indicate that the prevalence of DM has increased to 11.7% in China and 9.1% in the world, respectively. According to the latest report, the total number of adult diabetes have increased to 415 million by 2015.TCF7L2 gene polymorphism was associated with DN risk. In the subgroup analysis by ethnicities, the result indicated that TCF7L2 gene polymorphism was associated with DN risk in Asian and Caucasian, but not in Negroid. We presumed that the limited studies and sample size of subject might be the reason on this inconsistency. Rs7903146 mutation was associated with DM,38 and our subgroup analysis suggested that rs7903146 was also significant associated with DN. This was consistent with conclusions from FU et al32 and Buraczynska et al.21 Accordingly, we presumed that the rs7903146 mutation was important in the DM and DN, and the intervention of this mutation might bring new therapeutic strategy for DM and DN.Our result provided a new evidence that However, some limitations should be acknowledged in our meta-analysis. Firstly, DN is a continuous process which includes several different stages, from microalbuminuria to ESRD. Because of insufficient data, we did not probe subgroup analysis by different stages of DN, thus, we did not know whether TCF7L2 plays a different role in different stages, or how much influence our results. Secondly, the mechanism of DN is complicated; many factors may participate in its development, such as hypertension, poor glycemic control, diabetes duration, smoking, age, and others. Though we probed subgroup analysis, we still could not exclude all the potential influence factors. Thirdly, strong heterogeneities among the included studies made it less robust to interpret results. Finally, publication bias was found when assessing linear association between TCF7L2 polymorphism and DN risk.TCF7L2 gene polymorphism may contribute to the risk of DN, although whether the association is direct or indirect still cannot be determined. Anyhow, this finding provides a new point of view on the mechanism of DN, and could be a direction for future research.Notwithstanding the limitations discussed above, our analysis still includes most of the published literature referred to TCF7L2 and DN. The results provide new evidence on that


Heterotopic ossification (HO) is a frequent complication after total hip arthroplasty (THA). Nonsteroidal anti-inflammatory drugs (NSAIDs) have been used as routine prophylaxis for HO after THA. However, the efficacy of NSAIDs on HO, particularly selective NSAIDs versus nonselective NSAIDs, is uncertain.We searched PubMed, Embase, the Cochrane Central Register of Controlled Trials, and clinicaltrials.gov to identify randomized controlled trials with respect to HO after THA. Two reviewers extracted the data and estimated the risk of bias. For the ordered data, we followed the Bayesian framework to calculate the odds ratio (OR) with a 95% credible interval (CrI). For the dichotomous data, the OR and 95% confidence interval (CI) were calculated using Stata version 12.0. The subgroup analyses and the Grading of Recommendations, Assessment, Development and Evaluation (GRADE) approach were used.P = 0.046) more than a placebo. Selective NSAIDs could decrease DGSE  compared with the nonselective NSAIDs. There was no significant difference with respect to discontinuation caused by nongastrointestinal side effects (DNGSE) in NSAIDs versus a placebo  and in selective NSAIDs versus nonselective NSAIDs .A total of 1856 articles were identified, and 21 studies (5995 patients) were included. In the NSAIDs versus placebo analysis, NSAIDs could decrease the incidence of HO, according to the Brooker scale  and Delee scale . In the selective NSAIDs versus nonselective NSAIDs analysis, there was no significant difference  in the prevention of HO. NSAIDs could increase discontinuation caused by gastrointestinal side effects (DGSE) (OR = 1.28, 95% CI 1.00–1.63, NSAIDs might reduce the incidence of HO and increase DGSE in the short-term. Increased trials have demonstrated that nonsteroidal anti-inflammatory drugs (NSAIDs) are effective for the prevention of HO.12 Therefore, NSAIDs have been widely used for the prophylaxis of HO. However, the risk of gastrointestinal side effects caused by NSAIDs has drawn the attention of surgeons. Several studies showed that NSAIDs had a favorable influence on the prophylaxis of HO after THA, whereas other studies demonstrated that NSAIDs did not provide efficient prevention against the risk of HO.13 A previous Cochrane review14 investigated NSAIDs as prophylaxis for HO; however, that review had several defects. It included 2 quasi-randomized trials, applied an imprecise statistical analysis method, investigated NSAIDs versus a placebo rather than selective NSAIDs versus nonselective NSAIDs, and did not estimate the quality of evidence of every involved article. A recent meta-analysis15 investigated the difference between selective NSAIDs and nonselective NSAIDs as prophylaxis for HO after THA. This meta-analysis used an inexact statistical analysis method and did not evaluate the risk of bias and the quality of evidence of each involved article.Irradiation after surgery could decrease the incidence of HO.Because of the shortcomings of previous meta-analyses, their conclusions were not robust. Hence, we executed a Bayesian meta-analysis of randomized controlled trials to evaluate the efficacy and safety of NSAIDs for the prevention of HO after THA.We searched for the published results with respect to HO after THA in PubMed, Embase, the Cochrane Central Register of Controlled Trials, and clinicaltrials.gov. For this article, we searched databases by using the following terms (“Arthroplasty OR Replacement OR Hip” OR “arthropl∗ OR pros∗ OR surg∗ OR replac∗[tiab]” OR “hip prosthesis OR joint prosthesis”) AND “heterotopic ossification” AND (“pain [mesh]” OR “pain∗”). The initial search was limited to human subjects and RCTs regardless of the language of publication. In addition, we manually checked the reference lists of articles obtained and those from previously published systematic reviews and meta-analyses to identify the relevant reports. Ethical approval of this study was not necessary, as systematic review and meta-analyses do not involve patients.An article qualifying for this meta-analysis fulfilled the following inclusion criteria: participants assigned to undergo THA; treatment comparing NSAIDs versus the control or comparing two NSAID drugs; information assembled at the follow-up regarding HO, discontinuation caused by nongastrointestinal side effects (DNGSE) and discontinuation caused by gastrointestinal side effects (DGSE); and randomized controlled trials.Trials were excluded if they were abstracts, letters, case reports, case series, reviews, guidelines, nonhuman studies, or meeting proceedings concerning our investigation question; the studies used reduplicative data; or the patients had a definite contraindication for the treatment with an NSAID .Two reviewers independently extracted the following data from the qualified publication: the first author, year of publication, geographical location, span of the surgery, type and dose of the intervention and comparison, duration of the treatment, outcomes, duration of the follow-up, patient characteristics, number of patients, and study type. For information that could not been extracted from the involved studies, communication was made with the first author and study sponsors to obtain the relevant material. In the case of a disagreement not solved by discussion, a third reviewer judged the study.The primary outcome measure for this meta-analysis was radiographically determined HO. In the initial trials, there were several measurement scales, such as the Brooker scale, DeLee scale, Rosendahl scale, Hierton scale, and Hoikka scale, to grade the level of HO. If a certain classification method was used in only one trial, we described it in the narrative form. If the measurements of HO were made multiple times, the last outcome was recorded. The main secondary outcome measures were DNGSE and DGSE.16 using Review Manager, version5.3 . The content for the assessment consisted of the sequence generation (selection bias), allocation sequence concealment (selection bias), blinding of the participants and personnel (performance bias), blinding of the outcome assessment (detection bias), incomplete outcome data (attrition bias), selective outcome reporting (reporting bias), and other biases including whether the study was supported by external funding and whether the baseline was balanced. The judgment for each entry involved assessing the risk of bias as “low risk,” as “high risk,” or as “unclear risk,” which indicated a lack of information or uncertainty over the potential for bias. Two reviewers independently assessed each RCT, and any disagreements were resolved by discussion and consensus.Risk of bias assessment of each involved article was conducted in light of the Cochrane Handbook for Systematic Reviews of Interventions (version 5.1.0),17 to estimate the quality of the evidence. GRADE is a framework for illustrating the level of evidence. In this approach, the assessment was conducted based on 5 factors, including the risk of bias, inconsistency, indirectness, imprecision, and publication bias. We used GRADE Pro, version 3.6, to estimate the quality of evidence as high, moderate, low, or very low.We used the Grading of Recommendations, Assessment, Development and Evaluation (GRADE) approach18 We followed the Bayesian framework to explore the posterior distribution of HO. Noninformative priors were used to generate the outcomes. We estimated the odds ratio (OR) and 95% credible interval (CrI) by WinBUGS, version 1.4.3 . For the stratified model, in regard to HO, we operated 10,000 iterations and ran the burn-in of 1000 iterations. DNGSE and DGSE were conducted as the dichotomous outcomes, using Stata, version 12.0 . We calculated the pooled Peto OR estimates and 95% confidence interval (CI). The I2 test19 and chi-square test were used to assess the heterogeneity. When I2 > 50%, it indicated massive diversity between the studies, and a random-effect model was used. We used a funnel plot and Egger linear regression test to estimate the publication bias if the number of studies was larger than 10.20 Statistical significance was considered when the P value was less than 0.05. We executed the subgroup analyses by sample size and drug category to determine whether different sample sizes had effects on the estimates of DNGSE and DGSE and whether various drug categories yielded different influences on DNGSE and DGSE. We calculated their statistical power by using a commercially available software package . For the outcomes, a confidence level of 5% with a statistical power of 80% was used and regarded as being acceptable for medical purposes.21For the ordered outcome , we implemented the Bayesian random-effects meta-analysis according to the Bayesian framework model presented by Whitehead et al.30 Three of the studies were excluded because they were not studies on THA.33 One article was excluded because of repeated data.34 Finally, 21 articles fulfilling our inclusion criteria were included in our meta-analysis, with a total of 5995 participants. Figure We identified 1856 articles from PubMed, Embase, Cochrane Central Register of Controlled Trials, and clinicaltrials.gov, and the reference lists of the articles obtained (including those from previously published systematic reviews). Of these articles, 381 were removed due to duplicate reportage, and 1441 were excluded based on the titles and abstracts, which were case reports, reviews, irrelevant studies, systematic reviews, and meta-analyses. The remaining 34 articles were accessed for the full text and screened for further assessment, 9 of which were excluded because they were not randomized controlled trials.35 to 2649.13 Two articles were published in the 1980, 12 in the 1990, and 7 in the 21st Century. A total of 17 articles were from Europe, specifically, 2 from Finland, 4 from Denmark, 4 from Sweden, 3 from German, 1 from Belgium, 1 from France, 1 from The Netherlands, and 1 from Switzerland. Two trials were conducted in North America and Asia. Two studies originated in Oceania. One study was reported in Chinese, and the other studies were in English. Detailed information regarding the involved trials is shown in Table All of the included studies in our meta-analysis were randomized controlled trials. These studies were implemented from 1985 to 2011. The populations in the studies varied from 4338 of studies comprehensively described the generation of a randomized sequence, and the remaining studies did not demonstrate the randomization method. Blinding of participants and personnel was performed in 8 studies.40 The patients and healthcare teams were not blinded to treatment allocation in 1 study.37 There was insufficient information to permit judgment of “low risk” or “high risk” for the other studies. The outcome assessors did not have knowledge of the allocated interventions in 11 studies.42 All of the included articles displayed a low risk of bias for the incomplete outcomes and selective outcome reporting. Nine articles44 in the included studies displayed a low risk of bias for other biases, and three47 of the involved studies had a high risk of bias, with the remaining studies being indistinct.We used the risk of bias tool implemented in Review Manager 5.3 to evaluate the risk of bias in light of the Cochrane Handbook for Systematic Reviews of Interventions. The particular information of the risk of bias of the included articles is demonstrated in Figure 12 scale, there were 8 studies on NSAIDs as prophylaxis for HO after THA. NSAIDs showed a significant difference  in the prevention of HO compared with a placebo. A total of 5 studies related to NSAIDs versus a placebo contributed to the assessment of the Delee52 scale. Similarly, the NSAIDs significantly decreased the occurrence of HO after THA  compared with a placebo.Sixteen trials reported NSAIDs versus a placebo as prophylaxis against HO after THA. There were two classification methods  including more than 1 article. Based on the Brooker42 The 3 articles demonstrated that NSAIDs were more effective in preventing HO after THA than a placebo.Based on the Rosendahl, Hierton, and Hoikka scales, 3 articles were included.P = 0.297) . There was no publication bias found (Egger linear regression test: P = 0.325)   as well. No significant association was indicated . In the subgroup analyses relating to drug categories, the heterogeneity in nonselective NSAIDs was low  . Nonselective NSAIDs and selective NSAIDs demonstrated no significant association .Subgroup analyses with stratification by the sample size were performed. The heterogeneity in the articles with less than 100 samples was low  Figure , and the) Figure , and theP = 0.046) . No evidence of publication bias was found by Egger linear regression test (P = 0.885)   was low. There was no significant association . In the subgroup of drug categories, the trials that assigned selective NSAIDs had low heterogeneity   as well. No significant association was found between selective NSAIDs and DGSE (P = 0.236) and between nonselective NSAIDs and DGSE (P = 0.068).The subgroup analyses stratified by sample size showed that the heterogeneity in the articles with less than 100 samples  Figure , and the37 in Sudan using the Brooker scale, the second and the third level of the Brooker scale were combined. Therefore, the Bayesian framework could not be used in this study. The other 4 eligible articles were assessed. There was no significant difference  between the selective NSAIDs and nonselective NSAIDs as prophylaxis for HO after THA.Five studies were in the group of selective NSAIDs versus nonselective NSAIDs. However, in 1 studyP = 0.462) . The overall GRADE quality of evidence was judged to be high  Figure  comparedBecause there were no articles with less than 100 samples, subgroup analyses stratified by sample size could not be conducted.P = 0.042) . The overall GRADE quality of evidence was judged to be high  Figure . The staP = 0.628, I2 = 0.0%) . There was 1 article with less than 100 samples, and the heterogeneity could not be calculated.In the subgroup analyses with respect to sample size, the heterogeneity in the articles with equal to or more than 100 samples was low  Figure , and no The differences in 2  of the 3  analyzed variables were revealed in this meta-analysis of randomized controlled trials comparing NSAIDs with a placebo. According to the Brooker scale, NSAIDs could significantly reduce the incidence of HO after THA compared with a placebo. Regarding the Delee scale, NSAIDs could significantly decrease the incidence of HO after THA in comparison with a placebo. NSAIDs were not associated with DNGSE, whereas they increased the DGSE.50 and DGSE48 has been reported in individual trials or meta-analyses;54 however, they have drawn little attention. The American Academy of Orthopaedic Surgeons emphasizes preventing venous thromboembolic disease55 and diagnosing periprosthetic infections of hip prostheses56 in patients undergoing elective hip arthroplasty; however, prophylaxis for HO after THA has not been reported. We conducted a Bayesian meta-analysis based on randomized controlled trials comparing NSAIDs with a placebo to provide a more reliable proposal for the prevention of HO after THA.The overall incidence of HO14 in the overall incidence of HO; however, there were differences between our meta-analysis and that by Fransen and Neal.14 First, we excluded 2 quasi-randomized trials.58 Second, we introduced DNGSE, which was not analyzed by Fransen and Neal.14 By analyzing DGSE, we found that the risk of gastrointestinal side effects was increased 30.4% by NSAIDs, which was similar to the results of Fransen and Neal.14 However, Ma et al53 indicated that there was no significant difference in the gastrointestinal side effects between NSAIDs and a placebo. The number and quality of the included articles in his meta-analysis were worse than those in our meta-analysis.Our meta-analysis confirmed the results of Fransen and Neal13 concluded that NSAIDs had no effective influence on the prevention of HO after THA. First, compared with other studies primarily examining the efficacy of NSAIDs as prophylaxis for HO after THA, this trial investigated the efficacy of low dose of NSAIDs on thrombotic and hemorrhagic outcomes rather than the efficacy on the prevention of HO after THA. This situation might introduce clinical diversity so that different results occurred. Second, the poorer efficacy of aspirin compared with other NSAIDs might induce the difference.13 Finally, some outcome data on HO were not available because of the drop-out rate , which might also contribute to the diverse effect.In addition, a multi-center trial by Neal et alThe finding was observed by comparing selective NSAIDs with nonselective NSAIDs. There was no significant difference between the selective NSAIDs and the nonselective NSAIDs in the overall incidence of HO and DNGSE. Compared with nonselective NSAIDs, selective NSAIDs yielded a significant reduction in DGSE.51 A similar effect on the overall incidence of HO was observed in a recent meta-analysis of randomized trials.15 Our meta-analysis confirmed the results of Xue et al,15 and because of the use of Bayesian method, our outcomes were more robust.No statistically significant difference in the overall incidence of HO had been indicated in individual trials.15 It was inferred that DGSE was the major side effect of nonselective NSAIDs, including gastric perforation, gastric ulcer, and gastrorrhagia. Bombardier et al59 found selective COX-2 inhibitors might have a better manifestation in the protection of gastrointestinal functioning. This finding was consistent with our outcome.In addition, compared with the nonselective NSAIDs, there was a similar reduction of DGSE in selective NSAIDs between our meta-analysis and that by Xue et al.9 In this trial, the recruiting period was from 1996 to 1999. The time was so long that trial executors might select participants conforming to their will rather than the preplanned inclusion criteria.A previous study had illustrated that selective NSAIDs are less effective than nonselective NSAIDs for the prophylaxis of HO after THA.16 Therefore, our estimate outcome was more reliable and reasonable because we used the Bayesian framework.As the most extensive meta-analysis, all of the available randomized controls trials of NSAIDs versus a placebo and selective NSAIDs versus nonselective NSAIDs were included, and the intrinsic risk of bias in methodology and source of data in retrospective studies was overcome. Compared with previous reviews, our sample size was reinforced, and the credibility of demonstration was enhanced. We executed this meta-analysis based on Bayesian framework, which uses the Markov chain Monte Carlo method. A Bayesian framework could calculate a relatively exact distribution of the numbered sample, directly overcoming the disadvantage of traditional methods of making a statistical inference in light of asymptotic distribution of large sample, and took into account the model's uncertainty.Our meta-analysis has limitations. First, diverse dosages and types of NSAIDs were used in the 21 involved analyses. However, we did not investigate different dosages and types, because of the limited numbers of included studies, which might be a bias in the outcomes. Second, there were some patients undergoing revision surgery in the involved articles, and we could not extract the data of these patients to perform a subgroup analysis. Although there were few patients, the data might have influenced our outcomes. Finally, the statistical power in this study was slightly low, which indicated that potential differences were not identified.In our meta-analysis, the moderate quality of evidence indicated that NSAIDs could significantly yield a decrease in the incidence of HO, but increase DGSE. The high quality of evidence demonstrated that selective NSAIDs did not differ significantly from nonselective NSAIDs for the prevention of HO after THA, while produced fewer side effects. However, there was much indetermination in efficacy in long-range clinical outcomes. Given the side effects of nonselective NSAIDs, selective NSAIDs were recommended as prophylaxis for HO after THA. Even so, NSAIDs as prophylaxis for HO needs comprehensive evaluation in longer duration of follow-up and larger scale randomized controlled trials to identify the balance between efficacy and safety.