Previously reported findings on the association between folate intake or serum folate levels and esophageal cancer risk have been inconsistent. This study aims to summarize the evidence regarding these relationships using a dose-response meta-analysis approach. We performed electronic searches of the Pubmed, Medline and Cochrane Library electronic databases to identify studies examining the effect of folate on the risk of esophageal cancer. Ultimately, 19 studies were included in the meta-analysis. Summary odds ratios (ORs) were estimated using a random effects model. A linear regression analysis of the natural logarithm of the OR was carried out to assess the possible dose-response relationship between folate intake and esophageal cancer risk. The pooled ORs for esophageal cancer in the highest vs. lowest levels of dietary folate intake and serum folate were 0.63 (95% CI: 0.56-0.71) and 0.71 (95% CI: 0.55-0.92), respectively. The dose-response meta-analysis indicated that a 100 μg/day increment in dietary folate intake reduced the estimate risk of esophageal cancer by 12%. These findings suggest that dietary and serum folate exert a protective effect against esophageal carcinogenesis. Esophageal cancer is the most common upper gastrointestinal malignant tumor, ranking as the sixth most common cancer and the eighth leading cause of cancer deaths in the world . At presde novo thymidylate biosynthesis that will induce uracil mis-incorporation during DNA repair and synthesis resulting in DNA mutagenesis that in addition to DNA strand breaks and chromosomal damage triggers malignant transformation [Diet is an important modifiable factor that can modulate carcinogenesis. Among the dietary supplemental vitamins, folate shows promise in reducing the risk of carcinogenesis at some doses. Folate is a water-soluble B vitamin found in citrus fruits, green leafy vegetables, cruciferous vegetables and legumes, among others . Folate ormation . Althougormation –8. ThereFigure The main characteristics of the 19 selected studies were outlined in Table The quality of these studies was assessed by using NOS scale. The overall methodological quality of the studies is summarized in Table 2=70.2%; P<0.001) and the pooled OR of esophageal cancer for the highest vs. lowest level of dietary folate intake was 0.63 . Because only 4 studies reported serum folate levels, subgroup analysis based only on the geographic locations and histological types were performed for those .To study the relationship between dietary folate intake and the risk of esophageal cancer, dose-response of six case-control studies and one cohort study was analyzed. As shown in Figure P=0.29) and nonlinearity test (P=0.99) indicated that there is no linear or nonlinear relationship between the serum folate level and esophageal cancer risk.Further, four case-control studies were analyzed to find the dose-response relationship between the serum folate concentration and the risk of esophageal cancer. The results of both the linearity test .Publication bias was evaluated through visual inspection of funnel plots Figures  and 5. Tet al. and Bao et al. Excluding those studies reduced the heterogeneity, and the pooled OR of 0.542 (95% CI:0.329-0.92) was similar to the main finding.When a sensitivity analysis of dietary folate intake was conducted, entailing sequential exclusion of individual studies from the pooled analysis, the original conclusion remained unaffected Figure . SensitiAs of December 2014, 82 countries worldwide have passed legislations mandating folic acid fortification in at least one industrially milled cereal grain . AlthougThis meta-analysis, which included 19 studies, provides a quantitative estimate of the association of dietary folate intake and serum folate levels with esophageal cancer risk. Our analysis found that the dietary folate intake and the serum folate levels are inversely associated with the risk of esophageal cancer. Moreover, the dose-response analysis demonstrated that dietary folate intake was linearly associated with the risk of esophageal cancer. A 12% decrease in esophageal cancer risk was recorded for a 100 μg/day incremental increase in folate intake.The inverse relationship between dietary folate intake and esophageal cancer demonstrate in this study is consistent with a previous meta-analysis, which included nine case-control studies . However, our study is more reliable as it included a larger number of participants and derived stronger conclusions regarding the association between dietary folate intake or serum folate levels and esophageal cancer risk. Furthermore, our study is the first comprehensive meta-analysis of dose-response that quantitates the association between dietary folate intake and the risk of esophageal cancer.de novo, so they derive all the necessary folate from their diet [Since folate is a key modulator of DNA synthesis, repair and methylation, it was hypothesized to reduce cancer risk. Humans are unable to biosynthesize folate eir diet . Folate eir diet . Since teir diet  exploredeir diet . Howevereir diet . Animal eir diet , 40.In the subgroup analyses based on dietary assessment, histological type, source of control and study quality, we observed an inverse association between folate intake and esophageal cancer risk in all subgroups. However, in a sub-group analysis based on geographic locations, similar associations were absent in the Asian and American populations. The incidence and mortality rates of esophageal cancer showed wide geographic variation and there were marked differences between the high- and low-risk areas. People living in high esophageal cancer risk regions had significantly lower folate blood concentrations and nutrient intake than did those living in low risk regions . This maAlthough our study is the first comprehensive meta-analysis to evaluate serum folate levels and the risk of esophageal cancer, there were only four studies that could be analyzed. This may have limited the precision of the estimations of associations and reduced the credibility of the results. A larger-scale study with greater statistical power will be needed to better assess the associations, particularly for the subgroup analyses of interactions.Our study had several strengths that need to be highlighted. First, since our data included a broad folate intake range, it enabled accurate statistical analysis of the dose-response relationship between folate intake and the risk of esophageal cancer. Second, we assessed the methodological quality of the included studies by the Newcastle-Ottawa Scale (NOS). Third, our analysis provided an accurate assessment of the effect of folate levels upon esophageal cancer risk as we conducted subgroup analyses in specific populations. Further, high dietary folate intake and folate-rich foods may reflect a healthy lifestyle that includes other factors like never smoking, lower alcohol consumption and lower body weight that have been associated with a decreased risk of esophageal cancer. Therefore, factors such as age, sex, energy intake, alcohol use, smoking status, and treatment received were adjusted for in most of the studies included in this meta-analysis. Finally, sensitivity analyses of dietary folate studies showed that our results were robust, with no statistical evidence of publication bias.The limitations of our study were as follows: (1) since our search yielded only one prospective study, our results could be affected by recall bias and selection bias; (2) none of the studies analyzed total folate intake, comprising folate from the diet along with folate from supplements; (3) the analysis used pooled data , which prevented us from performing a more detailed analysis and obtaining more precise results.In conclusion, the results of this meta-analysis suggest that greater dietary intake of folate and higher serum folate levels are protective against esophageal cancer. And the dose-response analysis of dietary intake of folate show that every 100 μg/day increase of folate can reduced 12% risk of esophageal cancer. Our analysis also indicates that in future, well-designed large prospective cohort studies that repeatedly measure folate intake with long follow-up periods and adjustments for all potential confounders are necessary to accurately verify the association of dietary folate intake and serum folate levels with esophageal cancer.Relevant articles were identified by two reviewers through systematic searches of the Medline, Pubmed, and Cochrane Library electronic databases (from database inception to Jul 2016). The search was performed using the terms (“folate” OR “folic acid”) AND (“cancer” OR “neoplasm” OR “carcinoma”) AND (“cohort study” OR “case-control studies”). In addition, we scrutinized references from relevant original reports, review articles and meta-analyses to identify other pertinent studies. No language restrictions were imposed.A study was eligible for inclusion if the following criteria were met: (1) the study designed as a cohort, nested case-control or case-control study; (2) the study investigated the association between esophageal cancer and folate intake; and (3) the authors reported effect estimates  and 95% confidence intervals (CIs) for comparisons between high and low dietary folate intake or serum folate levels. When multiple levels of folate intake were presented, the ratio comparing the highest intake versus the lowest intake was chosen. In the case of duplicate studies, the most recently published study was chosen for inclusion.Data were extracted independently by two authors using a standard extraction form. The following data were extracted from each publication: the first author’s name, publication year, study design, geographic locations where the study was performed, type of controls in case-control studies, sample size (cases and controls or cohort size), measure and range of exposure, lowest folate level, highest folate level, difference between the highest and lowest folate levels, variables adjusted for in the analysis, and risk estimates with corresponding 95% CIs for the highest vs. lowest categories of folate intake or for each category. For studies that reported several multivariable adjusted-effect estimates, we selected the effect estimate that had been maximally adjusted for potential confounders.Study quality was assessed using the Newcastle-Ottawa quality assessment scale (NOS), which is a comprehensive tool that has been validated for evaluating the quality of observational studies in meta-analyses , 44. A N2 statistic, which is the proportion of total estimate variation attributable to study heterogeneity; I2 values of 25%, 50% and75% were used as cut-off points for low, moderate and high degrees of heterogeneity, respectively [We examined the association of folate intake and serum folate levels with the risk of esophageal cancer on the basis of the effect estimates (RR or OR) and 95% CI reported in each study. Heterogeneity was assessed using the Iectively , 48. We ectively , 50. We Lastly, we conducted a dose response analysis using the median or mean folate level and the adjusted natural log of the RRs or ORs with their standard error (SE). When the folate intake was reported by range, we assigned the midpoint of the upper and lower boundaries in each category as the average intake. When the highest category was open-ended, we considered the width of the category to be the same as that of the adjacent category. When the lowest category was open-ended, the lowest boundary was set to zero , 52. To


The current meta-analysis aims to evaluate the risk factors and neonatal outcomes of isolated Single Umbilical Artery (iSUA) in singleton pregnancy. Standard Mean Difference (SMD) or Weighted Mean Difference (WMD) was pooled for the maternal age, gravidity and parity, neonate birth weight and Apgar score one and five minutes after birth. We also pooled the odds ratios (ORs) at 95% confidence intervals (CIs) for maternal smoking status, the rate of neonate delivery before 37 or 34 weeks, Cesarean section (CS), the rate of being admitted to neonatal intensive care unit (NICU) and the serious adverse neonate outcome. Results show that maternal primigravidity  and female sex of the neonate  were associated with higher risks of iSUA. As compared to normal neonates, the neonates with iSUA had lower birth weight, worse Apgar score, increased risk of delivery before the normal gestational age, increased rate of CS due to fetal distress, increased rate of admission to NICU and prolonged NICU stay. However, no difference in neonatal mortality was observed. Maternal primigravidity and female neonate might associate with increased risk of iSUA. Identification of iSUA is of great importance for prenatal diagnosis and may improve neonatal outcomes. This malformation has a reported incidence of up to 1% of pregnancies. The SUA incidence is 3–4 times higher in multiple pregnancies than in singleton pregnancy2. The most widely accepted causes for this anomaly are primary agenesis and/or later thrombotic atrophy3.The normal umbilical cord contains two arteries and one vein (three-vessel cord). Single Umbilical Artery (SUA) is characterized by the absence of either the left or right umbilical artery5, maternal smoking status6, ethnicity4, multiparity and multiple gestation7. However, the results were inconsistent, possibly due to the limited sample size, varying geographic regions or differences among research subjects. For example, a study from Tulek et al. suggested that neither maternal age nor preterm delivery were associated with the development of iSUA8. Most cases of SUA were not associated with the presence of other malformations, in which chromosomal alterations were usually not present. However, certain obstetric complications were often observed in iSUA cases, such as fetal growth restriction and increased perinatal mortality10. Clinical relevance of iSUA was evaluated after birth in a prospective cohort study by Chetty-John et al.4. No significant difference in physical growth or neurological development was observed4. There was also no evidence of increased need for admission to neonatal intensive care units11. Leung and Robson12 noted that neonatal low birth weight was observed in around 28% of all SUA cases. Burshtein et al.10 reported that iSUA in fetus was a risk factor of Apgar score, whereas no relation was observed between the fetal iSUA and Apgar score in other studies8. Neonates with SUA and iSUA had increased rates of prematurity, growth restriction and adverse neonatal outcomes1. Studies also indicated that risks of some adverse neonatal outcomes, such as low birth weight, preterm delivery and low Apgar score, were increased in cases with SUA14.Previous studies have suggested many risk factors for iSUA, such as maternal ageIn this study, we performed a meta-analysis to clarify the association of iSUA with the neonatal outcomes and the associated risk factors.A total of 57 relevant studies were retrieved from PUBMED, EMBASE and MEDLINE. We excluded the books, reviews and studies which are unrelated to humans, studies in which the control groups were not normal umbilical cord fetuses, studies in which case groups were not isolated SUA and studies of which the raw data could not be retrieved. For overlapped studies, only the ones with the most extensive results were included. After carefully examining all the literatures, a total of 12 qualified studies was included for the final meta-analysis. The data collection flow chart is shown in Fig. 2 = 0%). As to fetal risk factors, female fetuses were associated with an increased risk for SUA with the overall OR of 0.805 . As shown in Table The results from different analysis methods for evaluating the risk factors of isolated SUA are summarized in Table Statistically significant association was detected between iSUA and decreased neonatal birth weight or Apgar score at 1 min postpartum in our meta-analysis. In comparison with neonates with normal umbilical vessels, the iSUA neonates had lower birth weight  Fig. , Table 4Our study also reveals that iSUA was a risk factor for CS because of fetal distress [OR: 2.258 , Fig. The funnel plot, Begg’s test and Egger’s test were utilized to assess the potential publication bias of included studies. The results indicated no evidence of publication bias for all the subgroup analyses Tables , 3 and 415. Over the past 30 years, numerous maternal and fetal risk factors were reported to be associated with SUA8. However, due to small sample size and difference in research subjects, previous studies provided limited information and arrived at controversial conclusions. Therefore, a systematic meta-analysis of all available qualified studies may provide us with definitive answers.SUA is an abnormal condition of the umbilical cord in which one artery is missing. The prevalence of SUA ranges from 0.2% to 11%16. The results suggested that there was no significant association of iSUA with fetal growth, perinatal mortality or aneuploidy. A large-scale prospective cohort study is needed to reach definitive conclusions on the appropriate work-up in iSUA pregnancies. Thus, we collected a large number of qualified studies for a meta-analysis, aiming to evaluate the risk factors and neonates outcomes of iSUA in singleton pregnancy.Meta-analysis is a powerful approach for investigating the risk factors and neonatal outcomes of iSUA in singleton pregnancy. To date, only 1 meta-analysis publication has focused on the association among iSUA, fetal growth, aneuploidy and perinatal mortality14 that suggested that iSUA was associated not only with anomalies at birth and but also with increased risk of adverse pregnancy outcomes. In our meta-analysis, we found that iSUA neonates had a higher rate of preterm birth < 37 weeks, which suggested that iSUA might be a risk factor for premature birth. Similar results have been published in a previous report6. However, our results did not suggest that iSUA is a risk factor for neonatal mortality, which is different from the finding in a previous study in which iSUA was shown to be associated with perinatal mortality. This might be due to a much larger sample size in the current meta-analysis. It should be noted that a meta-analysis17 study on a similar issue was recently published after the submission of our manuscript. This study was focused on evaluating the association of iSUA with pregnancy outcomes and perinatal outcomes. No association between iSUA and pregnancy outcomes was identified. iSUA was found to be correlated only with some perinatal complications in this study17.In the current study, we revealed that both maternal status and neonatal sex might be risk factors for iSUA. iSUA itself is a risk factor for Cesarean section and iSUA neonates might have prolonged NICU stay. This result is consistent with the findings of previous reportsDespite the clinical significance of our study, there are still some limitations. First, selection bias might exist as all studies included in this meta-analysis were published in English. Studies in languages other than English that may have impact on the evaluation were excluded. Second, although all cases and controls in each study were well-defined following the inclusion criteria, there might be factors that were not taken into account but might influence our results if included. Future analyses including more studies on these risk factors are needed to further confirm our findings.In this article, we presented a meta-analysis to evaluate the neonatal outcomes and possible risk factors associated with iSUA. Our results suggest that maternal primivalidity and the female sex of neonates might be risk factors for iSUA. Fetuses and neonates with SUA and/or iSUA have increased risk of adverse outcomes. Therefore, the diagnosis of iSUA is necessary during pregnancy and attention should be paid to adverse outcomes associated with iSUA for neonates. As such, surveillance with iSUA would improve neonatal outcomes.For the first-round exclusion, articles were searched in the NCBI Global Cross-database, including PUBMED, EMBASE and MEDLINE, using “single umbilical artery” or “two umbilical vessels” or “SUA” and  and  as key words. Where appropriate, Standard Mean Difference (SMD) or Weighted Mean Difference (WMD) were pooled for the maternal age, gravidity and parity, neonatal birth weight and Apgar score at both one and five minutes postpartum, and odds ratios (ORs) for maternal smoking status, the rate of neonatal delivery before 37 weeks or 34 weeks, Cesarean section (CS) because of fetal distress, the rate of admission to NICU, and the serious adverse neonatal outcome .4. SUA was considered to be isolated if there were no additional structural anomalies and markers for an euploidy or small for gestational age (SGA) at the time of the ultrasound scan16. Studies were excluded if iSUA was not diagnosed at birth or if the study was only on twin pregnancies. Other exclusion criteria were: presence of any fetal malformation, marker of euploidy by ultrasound examination, chromosomal abnormality established by fetal karyotyping or multi-fetal pregnancy, or absence of a SUA umbilical cord at delivery or by pathological examination9.Retrospective cohort studies and retrospective case-control studies were considered eligible if the iSUA described in the study was identified by ultrasound in single pregnancy. In other words, our meta-analysis included all studies on singletons with at least 24 weeks’ gestation and SUA at birth but without identifiable congenital anomalies11. Composite adverse outcomes included the following: Cesarean section (CS) because of fetal distress, prolonged neonatal admission and low Apgar score8. Neonatal mortality was considered as the serious adverse event.Possible risk factors included maternal age, gravidity, parity, smoking status, the BMI and neonatal sex. Neonatal data included birth weight, Apgar scores, mode of delivery, rate of preterm delivery , NICU admissions, and neonatal admission days (length of stay)All studies that are related to the association between the iSUA and neonate’s outcomes or possible risk factors were included. We collected the data of outcomes from all qualified studies for the construction of a 2 × 2 (dichotomous data) or 3 × 2 (continuous date) table.2 index. When I2 index was less than 50%, the Mantel-Haenszel (M-H) fixed-effects model was adopted for dichotomous data and Inverse-Variance (I-V) fixed-effects model was used for continuous data. Otherwise, the DerSimonian and Laird (D-L) random-effects model was applied for both dichotomous data and continuous data. A forest plot and Begg’s funnel plot were generated for evaluating publication bias. Higher asymmetry in the funnel plot indicates potential publication biases. To further assess publication bias, Egger’s test was employed for each dataset.STATA 12 software was used for all analysis. The overall OR and corresponding 95% CI were calculated for dichotomous data, while the standardized mean difference (SMD) or weighted mean difference (WMD) and 95% CI were calculated for the continuous data. Heterogeneity was measured using the ISupplementary Info File


Effects sizes showed a substantial heterogeneity. As a result, the 95% prediction intervals were wide and ranged between a large negative and a large positive effect size for odor threshold, , and for odor identification, . Exploratory analyses suggested that age and IQ may be potential moderators. To conclude, the large heterogeneity is consistent with the notion of both hyposensitivity and hypersensitivity in individuals with ASD. However, future research needs to predict and test the specific direction of the effect to provide convincing evidence for atypical olfactory functions in ASD.Autism Spectrum Disorders (ASD) are often accompanied by atypical visual, auditory, and tactile sensory behavior. Evidence also suggests alterations of the olfactory system, but the pattern of findings appears mixed. To quantify this pattern systematically, we conducted a meta-analysis. Studies were included if they examined olfactory function  in ASD compared with healthy age-matched control groups. We also coded for the potential moderators gender, age, and IQ. Articles were identified through computerized literature search using Web of Science, PubMed, and Scopus databases. A total of 11 articles compared odor threshold and/or odor identification between cases and controls (for threshold,  Autism Spectrum Disorders (ASD) are characterized by social and communication difficulties, alongside repetitive behaviors and special interests . In addiAvailable evidence also suggests alterations of the olfactory system in ASD, but the pattern of findings is mixed. Inconsistent observations are reported for both sensory-driven olfactory tasks  and higher-order olfactory functions . Although some studies reported either enhanced olfactory sensitivity in ASD  or impaiSome degree of heterogeneity may be expected because previous studies used different methods, had small sample sizes, and may have sampled from different populations . However, to determine if there is actual heterogeneity among the effect sizes that is not simply due to chance, we conducted a meta-analysis of existing studies examining olfactory function in individuals with ASD. We examined olfactory threshold and odor identification in ASD as compared with healthy age-matched control groups. Further, in an exploratory meta-regression, we investigated the potentially moderating role of available variables  upon the observed olfactory differences between ASD and controls.∗, ASD and olfaction∗, ASC and olfaction∗, HFA and olfaction∗, and AS and olfaction∗. The search was limited to English language articles. In addition, a manual review of articles was performed using cross-references from the original articles and reviews.Published studies were identified through computerized literature searches in Web of Science, PubMed, and Scopus databases for relevant studies targeting human olfaction in ASD, ASC (Autism Spectrum Condition), HFA (High-Functioning Autism), and AS (Asperger Syndrome). The search keywords were: autism and olfactionFigure 7). Based on these criteria, two of the authors (ML and CT) and one research assistant independently reviewed and extracted data from each potential study. Disagreements were resolved via discussion of all authors until a consensus decision was reached. A total of 11 studies were included (see Tables 2), and 15 studies were excluded (Table 3).Studies that were included in the meta-analysis focused on standard or experimental tasks of olfactory function in ASD, and had an age-matched control group of healthy participants without ASD , we used the reported inferential statistics to compute Hedges’ g. Hedges’ g and its estimated variance were computed (R toolbox meta (version 4.6-0) to perform the meta-analysis  and OSF (10.17605/OSF.IO/V24NE), as recommended , and sample size for each group. Because the studies used different measures, we computed the standardized, unbiased effect size estimate Hedges’ g . If the computed  and thenanalysis . Becauseysis  = 34.73, p < 0.0001, T = 0.70, and about 83% of the observed variance reflected actual differences in the effect sizes, I2 = 82.7, 95% CI . The predicted effect size of a future study ranges between a large negative and a large positive effect size, as shown by the 95% prediction interval . Consistent with this heterogeneity, the 95% CI of the mean estimate of the random-effects model was wide and overlapped zero, . As shown in Figure 2, the heterogeneity is also illustrated in the contour-enhanced funnel plot  = 38.87, p < 0.0001, T = 0.77, and about 82% of the observed variance reflected actual differences in the effect sizes, I2 = 81.9, 95% CI . The prediction interval ranged between a large negative and a large positive effect size, . Consistent with this heterogeneity, the 95% CI of the mean estimate of the random-effects model was wide and overlapped zero, . The heterogeneity is also illustrated in the contour-enhanced funnel plot . Most studies had low precision and varied widely. The figure suggests that there was no particularly bias toward significant findings.A similar pattern of results was obtained for the identification task. g = 1.99), the above analysis was repeated without this study . Results were unaffected: the large heterogeneity remained, Q(df = 6) = 26.94, p = 0.0001, T = 0.65, I2 = 77.7, 95% CI , and 95% prediction interval, . Consistent with this heterogeneity, the 95% CI of the mean estimate of the random-effects model was wide and overlapped zero, , although it was centered closer to zero than in the analysis of all studies.Because for the identification task, one study  had a laTaken together, results showed that effect sizes varied substantially among the studies on both threshold and identification tasks. These results are consistent with the notion that ASD may lead to both hyposensitivity and hypersensitivity on these tasks. To identify potential moderators of the direction of this effect, we explored the contributions of three variables that we could extract from the studies in explaining the variability among effect sizes: proportion of males, mean age, and mean IQ score in the ASD group.Figure 5), proportion of males showed no clear relationship with sensitivity, B = 0.01, 95% CI , R2 < 0.01. In contrast, mean age and mean IQ tended to be associated with the performance difference between ASD and control group. For mean age, B = -0.05, 95% CI , R2 = 0.80; and for mean IQ, B = -0.10, 95% CI , R2 = 1.00. As shown in Figure 5, for younger participants (below 30), the ASD group (vs. the control group) showed hyposensitivity, whereas for older participants (above 35), the ASD group showed hypersensitivity. Similarly, for participants with IQs below 113, the ASD group showed hyposensitivity, whereas for participants with IQs above 113, the ASD group showed hypersensitivity. However, mean age and mean IQ correlated with each other (r = 0.72). Given the small number of studies, the unique contribution of each variable cannot be estimated.Results of these exploratory meta-regressions showed that for threshold , proportion of males and mean age showed no clear relationship with sensitivity; for proportion of males, B = 0.02, 95% CI , R2 = 0.03; and for mean age, B = 0.03, 95% CI , R2 < 0.01. In contrast, mean IQ tended to be associated with the performance difference between ASD and control group; for mean IQ, B = 0.10, 95% CI , R2 = 0.45. As shown in Figure 6, with increases in mean IQ, the ASD group increased in hyposensitivity.For identification  for the threshold task  and for the identification task . These intervals imply that for a future study, the true effect size may fall between very negative and very positive. As such, the results are consistent with the idea of both hyposensitivity and hypersensitivity in individuals with ASD.The goal of the present study was to summarize quantitatively the empirical evidence for changes in olfactory function in individuals with ASD. Although only few studies with small sample sizes were available, the meta-analysis provided strong evidence that the effect sizes varied substantially among the studies for both threshold and identification tasks. This heterogeneity is clearly apparent in the large 95% prediction intervals (see r = 0.72). This makes it difficult to separate the unique contribution of each variable. For identification, increases in mean IQ were associated with increased hyposensitivity. These results are surprising because they suggest that mean IQ is associated with opposite effects depending on the task: increased mean IQ is associated with hypersensitivity on a threshold task and hyposensitivity on an identification task. However, these findings from the meta-regression need to be interpreted carefully , and subsequent analyses could evaluate the robustness of the findings depending on the prior . Notably, the present results imply that 35-year old males with ASD and an IQ around 115 would show opposite effects on threshold and identification tasks, that is, hypersensitivity on the threshold task and hyposensitivity on the identification task. To determine if this is not simply a chance finding, future research should study well-characterized individuals with standardized methods and consider the role of potential moderators. Any potential moderators should be reported completely to avoid missing data. Sample size should be large to reduce the uncertainty in the obtained estimate. Although a power analysis is often recommended to determine sample size, a power analysis seems cumbersome in the present context, as the effect sizes were heterogeneous and their direction was unclear. Critically, a power analysis is meaningful only from the perspective of null hypothesis significance testing with its focus on the ticipant . This sihe prior . Anotherhe prior ; for an he prior . Also, Bhe prior . These che prior . Furtherhe prior . Last, the prior . In parthe prior .The present meta-analysis provides convincing evidence for heterogeneity among effect sizes that exceeds that expected by chance. Demographic variables such as age and IQ may account, at least in part, for this heterogeneity, and their effects may differ depending on the task. Future studies should be conducted on well-defined samples with standard methods to assess olfactory functions in ASD. If a study is pre-registered and uses Bayesian statistics, participants could be added consecutively until the evidence is convincing for hyposensitivity, hypersensitivity, or neither.ML and SW developed the study concept. ML and CT jointly searched the databases and extracted the data. SW analyzed the data and produced the figures. ML, CT, and SW wrote the 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.


CD44 have been shown to be associated with susceptibility to various cancers; however, the results are confounding. Hence, we performed a meta-analysis to clarify these associations more accurately. Overall, rs13347  and rs187115  polymorphisms were found to significantly increase the cancer risk in Asians. On the other hand, rs11821102 was found to confer low risk . Based on our analysis, we suggest significant role of CD44 variants  in modulating individual's cancer susceptibility in Asians. Therefore, these variants may be used as predictive genetic biomarkers for cancer predisposition in Asian populations. However, more comprehensive studies involving other cancers and/or populations, haplotypes, gene-gene and gene-environment interactions are necessary to delineate the role of these variants in conferring cancer risk.CD44 is a well-established cancer stem cell marker playing a crucial role in tumor metastasis, recurrence and chemo-resistance. Genetic variants of  Cancer, which is an extremely complex and multifaceted disease involving multiple steps, is a leading cause of death worldwide. During last decades, considerable advancements have taken place in the development of better therapeutic interventions for cancer; however, chemo-resistance and disease recurrence have resulted in minimal disease outcome and poor survival rates . Cancer CD44 to generate many standard (CD44s) and variant isoforms (CD44v) of varying sizes [Cluster of differentiation (CD) 44, a well-recognized CSC marker , 9, is ang sizes , 13. Altng sizes -17. In ang sizes -20.Several evidences have firmly established the role of CD44 in cell differentiation, epithelial -mesenchymal transition (EMT), invasion and metastatic spread in various human cancers -25. In aCD44 in carcinogenesis, several studies have explored the role of genetic variants of CD44 in cancer susceptibility, prognosis and chemotherapeutic response in various human cancers [CD44 gene  and cancer risk.Considering the important role of  cancers -38. HoweCD44 polymorphisms  with cancer susceptibility [According to the search strategies mentioned above, we found a total of 13 case-control studies investigating the association of tibility -49. HoweAccording to the Newcastle-Ottawa quality assessment scale (NOS), the quality of all recruited case-control studies and their total quality scores are summarized in Table CD44 rs13347 meta-analysis, a total of 12 articles [vs. C: OR = 1.30, 95% CI = 1.09-1.56, p = <0.004; CT vs. CC: OR = 1.29, 95% CI = 1.05-1.58, p = 0.015; TT vs. CC: OR = 1.77, 95% CI = 1.28-2.44, p = <0.000, CT+TT vs. CC: OR = 1.34, 95% CI = 1.08-1.67, p = <0.009, Table vs. CC: ph = 0.085, I2 = 46.057; TT vs. CC: ph = 0.288, I2 = 1o.17) while removal of above studies together with Verma et al. [vs. C: ph = 0.576, I2 = 0.000; CT+TT vs. CC: ph = 0.386, I2 = 4.764) However, it was found to significantly change the pooled results. In our sensitivity analysis, we did not find any obvious change in the corresponding pooled ORs after removing one study each time for a genetic model, thereby confirming reliability of our results.For articles , 38-48 wvs. C: OR = 1.76, 95% CI = 1.50-2.07, p = <0.000; CT vs. CC: OR = 1.80, 95% CI = 1.64-1.97, p = 0.000; TT vs. CC: OR = 2.58, 95% CI = 1.93-3.45, p = <0.000, CT+TT vs. CC: OR = 1.94, 95% CI = 1.66-2.27, p = <0.000, Figure vs. C: OR = 1.81, 95% CI = 1.56-2.09, p = <0.000; CT vs. CC: OR = 1.81, 95% CI = 1.66-1.98, p = 0.000; TT vs. CC: OR = 2.79, 95% CI = 2.09-3.74, p = <0.000, CT+TT vs. CC: OR = 1.99, 95% CI = 1.72-2.30, p = <0.000, Figure vs. C: OR = 1.70, 95% CI = 1.14-2.55, p = <0.010; CT vs. CC: OR = 1.75, 95% CI = 1.18-2.59, p = 0.005; TT vs. CC: OR = 2.73, 95% CI = 1.20-6.23, p = <0.017, CT+TT vs. CC: OR = 1.89, 95% CI = 1.19-2.99, p = <0.007, Table In subgroup analysis based on study design and genotyping method (Taqman and/or other), the significant association was limited only to population based studies . This may be because the small sample size of the subgroup did not possess sufficient statistical power to detect a weak effect.Further, in stratified analysis this association was lost in each subgroup except for HNC at allele level as well as at hetero genotype model  investigated the association of rs10836347 polymorphism in various cancers , 46, 48.For rs713330, we found a total of six eligible studies with 3453 cancer cases and 3958 controls , 48. Thevs. AA: OR = 2.34, 95% CI = 1.67-3.27, p = <0.000 and AG vs. AA: OR = 1.59, 95% CI = 1.37-1.84, p = <0.000, Table vs. A allele , as well as in dominant models  and dominant model . Removal of study by Sharma et al. [vs. A: ph = 0.837, I2 = 0.000; AG+GG vs. AA: ph = 0.656, I2 = 0.000. Due to limited number of studies, we could not perform stratified analysis for this SNP. Further, our sensitivity analysis confirmed the robustness of our findings.For rs187115, only six studies investigated the cancer risk, however the study of Verma et al.  deviateda et al.  was founFor rs13347, the review of funnel plot showed slight apparent asymmetry. Further, Egger's test as well as Begg and Mazumdar rank correlation tests did not demonstrate significant asymmetry except in the heterogenotype as well as dominant models  of C allele , 48 and C allele , 48. In C allele . The rs1C allele , 40, AMLC allele  and breaC allele , and thiC allele , 48. AltC allele , 43-46, CD44. Intronic SNPs have been shown to play an important role in gene function by regulating its transcription and splicing [CD44 rs187115 variant allele were associated with increased bone metastasis and tumor stage, as well as with decreased survival rate in NSCLC patients. Thus, this variant was suggested as a potential predictive marker of survival in NSCLC patients [The rs187115 SNP is located in the first intron of splicing . Previousplicing . Furthersplicing , 46, 53.patients .CD44 deregulation. Further, our in-silico analysis also revealed the role of CD44 rs1182102 in transcriptional regulation  methodologic quality. In addition, our study has improved the statistical power of the analysis since we pooled large number of cases and controls from various studies. We also performed sensitivity analysis and multiple corrections to remove any false result, though the result remained unaffected, thereby adding weight to our findings. Since cancer is a highly fatal disease, our results investigating the association of functional SNP in CD44 gene may have clinical significance in that they can help to identify interindividual differences in tumor susceptibility, recurrence capacity and chemoresistance among patients. However, care should be taken to interpret these results with caution as overall our study indicate moderate or weak evidence for association mainly due to limited number of studies.To the best of our knowledge, we are the first to perform such a comprehensive meta-analysis of common functional polymorphisms of the Though, we have collected all published articles till date, we could not perform a comprehensive subgroup analysis as the number of available studies were limited to Asian population and also for limited cancer types. In addition, there was significant heterogeneity for rs13347 meta-analysis, although in subgroup analysis it was removed or decreased Further, our results are based on unadjusted or crude estimates and the roles of haplotypes, gene-gene, and gene-environment interactions, as well as linkage disequilibriums were not considered. Last but not the least, we could not exclude the possibility of selection bias as study selection was limited to published results, articles in English language only and methodologies using different genotyping methods and study designs.CD44 SNPs in modulation of cancer risk. Specifically, rs13347 and rs187115 may be used as potential biomarkers for cancers in Asian populations. However, further analysis considering the aforementioned limitations and prognostic significance of CD44 are required to better understand the role of these CD44 SNPs in cancer risk.We demonstrated a significant association of CD44 gene”, “Cluster of differentiation”, in combination with “single nucleotide polymorphism (SNP) /variation/genotype”, and “cancer/carcinoma” or “tumor”. All published case-control studies investigating the association of CD44 gene polymorphisms with human cancer susceptibility in English language were searched until May2016. All relevant studies were collected after thorough investigation of the abstracts of potential articles. Further, the reference lists of the selected articles and related reviews on the topic were manually examined to collect additional relevant studies.Following the PRISMA statement , we perfCD44 polymorphism with cancer risk having sufficient information to calculate the relative risk and 95% confidence intervals (CI), histo-pathologically confirmed cancer cases and healthy controls . On the other hand, studies unrelated to cancer research or lacking control population or sufficient data, and those not in accordance with Hardy-Weinberg equilibrium (HWE) in control groups were excluded from the meta-analysis. Duplicate or ecological studies, case reports, reviews, abstracts, comments and editorials were also excluded from the present meta-analysis.The selection criteria of the studies were; original case-control study examining the association of Two independent investigators separately weighed the eligibility of each study according to the inclusion and exclusion criteria listed above and any disagreements were further resolved by discussions and agreements. Data such as first author name, publication year, country of origin, ethnicity, genotyping methods, cancer types, frequency of cases and controls, genotype frequencies, minor allele frequencies, etc., were cautiously extracted from all eligible studies.The quality of each studies included in this meta-analysis (Ref) was rigorously evaluated independently by two authors (Rai Rajani and Gupta Usha), by using the Newcastle-Ottawa quality assessment scale (NOS) , 56 and 2 value and Chi-square-based Q statistics (significant at p < 0.05). I2 = 0%, 25%, 50% and 75% were considered as no, low, moderate, and high observed heterogeneity, respectively [corr. A pcorrvalue less than 0.05 was considered as significant [http://www.tufts.edu/~mcourt01/Documents/Court%20lab%20-%20HW%20calculator.xls). Further, in-silico study was performed by using online Web servers- FastsnP (http://fastsnp.ibms.sinica.edu.tw) and F-SNP (http://compbio.cs.queensu.ca/F-snP/) to predict the functional effect of each SNPs.All statistical analyses were conducted using the Comprehensive Meta-analysis software . The pooled ORs were estimated for allele contrast, log-additive and dominant models. Odd's ratio greater than 1 is considered significant. Heterogeneity was measured using the Iectively . In the ectively , 59. Funectively . Moreovenificant . Hardy-WThe credibility of the cumulative association of CD44 polymorphisms and the cancer risk was scrutinized by using Venice interim criteria  includin


Achilles tendinopathy is the most prevalent tendon disorder in people engaged in running and jumping sports. Aetiology of Achilles tendinopathy is complex and requires comprehensive research of contributing risk factors. There is relatively little research focussing on potential biomedical risk factors for Achilles tendinopathy. The purpose of this systematic review is to identify studies and summarise current knowledge of biomedical risk factors of Achilles tendinopathy in physically active people.Research databases were searched for relevant articles followed by assessment in accordance with PRISMA statement and standards of Cochrane collaboration. Levels of evidence and quality assessment designation were implemented in accordance with OCEBM levels of evidence and Newcastle-Ottawa Quality Assessment Scale, respectively.A systematic review of the literature identified 22 suitable articles. All included studies had moderate level of evidence (2b) with the Newcastle-Ottawa score varying between 6 and 9. The majority (17) investigated genetic polymorphisms involved in tendon structure and homeostasis and apoptosis and inflammation pathways.COL5A1 genetic variants were the most extensively studied, particularly in association with genetic variants in the genes involved in regulation of cell-matrix interaction in tendon and matrix homeostasis. It is important to investigate connections and pathways whose interactions might be disrupted and therefore alter collagen structure and lead to the development of pathology. Polymorphisms in genes involved in apoptosis and inflammation, and Achilles tendinopathy did not show strong association and, however, should be considered for further investigation.Overweight as a risk factor of Achilles tendinopathy was described in five included studies that investigated non-genetic factors. This systematic review suggests that biomedical risk factors are an important consideration in the future study of propensity to the development of Achilles tendinopathy. The presence of certain medical comorbidities and genetic markers should be considered when contemplating the aetiology of Achilles tendinopathy. Further elucidation of biomedical risk factors will aid in the understanding of tendon pathology and patient risk, thereby informing prevention and management strategies for Achilles tendinopathy.CRD42016036558PROSPERO  Increased BMI and adverse lipid profile may be important biomarkers of Achilles tendinopathy.Further research is required to confirm an association between genetic variation, of genes encoding collagen proteins and proteins involved in pathways of tendon homeostasis and Achilles tendinopathy.Genetic risk factors of Achilles tendinopathy may be modified by geographic factors.Achilles tendinopathy is one of the most prevalent overuse tendon injuries associated with physical activities such as running and jumping . It is tOriginally, the pain of Achilles tendinopathy was attributed to an inflammatory process. While inflammatory cells have been observed, particularly in the early stages of Achilles tendinopathy, the response does not seem to be a traditional inflammatory response , 7. SeveIntrinsic and extrinsic aetiological factors interact in the genesis of Achilles tendinopathy. Intrinsic risk factors include demographic factors  and genetic polymorphisms; and local anatomical factors include leg length discrepancy, malalignment and decreased flexibility. Extrinsic factors comprise therapeutic agents , environmental conditions, and physical activity-related factors, including training patterns, technique and equipment , 2. The The contribution of biomedical risk factors to the development of Achilles tendinopathy remains unclear. These may include medical comorbidities and physiological, biochemical and genetic factors. Imaging and histopathological analysis of the degenerative and recovery processes in tendon have provided a better understanding of the processes associated with tendon structure and metabolism. In terms of biochemical and metabolic processes, tendon tissue is relatively inert, with new collagen synthesis being a slow process . LookingThere is relatively little research focussing on potential biomedical risk factors for Achilles tendinopathy. The purpose of this systematic review is to identify studies and summarise current knowledge of biomedical risk factors of Achilles tendinopathy in physically active people. This information will contribute to the understanding of Achilles tendinopathy and may inform future prevention strategies.This systematic review was registered in the international prospective register of systematic reviews PROSPERO (registration number CRD42016036558). Systematic review methodology included a thorough search of all accessible databases of scientific articles. Keywords were selected by following PICO  principle . Four daSelected articles from each database were combined in one electronic library EndNote . Duplicates were removed automatically in the electronic library. The titles and abstracts were then scanned, and clearly irrelevant studies and the remainder of the duplicates were removed. Following exclusion/inclusion criteria, two reviewers examined the abstracts of the remaining articles and selected studies for the full-text review. Inclusion criteria were the following: (a) retrospective and prospective studies, case-control studies and cohort studies investigating risk factors of Achilles tendinopathy; (b) clear definition of the studied injury and medical diagnosis of the injury; and (c) physically active people as studied subjects. Discrepancies in the lists of selected articles were resolved through the discussion between the reviewers.The quality of the included studies was evaluated applying the Newcastle-Ottawa Quality Assessment Scale (NOS) for case-control and cohort studies . This asOxford Centre for Evidence-based Medicine (OCEBM) levels of evidence was used as a guidance to evaluate selected articles. Levels of evidence could be identified according to the type of clinical question .Results of the studies are described in narrative form, and odds ratios and confidence intervals or relative risks of developing Achilles tendinopathy are reported in Table A total of 1631 articles were retrieved after the search of four databases. Following removal of duplicates, 1185 articles remained. Following screening of titles and abstracts, reviewers selected 30 articles as potentially relevant. The final 22 articles were included in the review after full-text screening and discussion between the reviewers. A flow chart of the article selection is given in Fig. The quality assessment using the NOS tool is presented in the Table The selected studies focused on a diverse range of biomedical risk factors for Achilles tendinopathy such as demographic factors, chronic medical conditions, lifestyle habits and genetic factors, including those contributing to collagen structure and tendon homeostasis and those involved in apoptosis and inflammation. The majority of the included studies were case-control studies Fig. . The maiBiomedical risk factors include non-modifiable factors such as age, sex, weight and chronic medical conditions. A cross-sectional study on 178 master track and field athletes did not find any association between Achilles tendinopathy (as determined by VISA-A scores) and age, sex, weight or height . HoweverA study of 450 cases of Achilles tendinopathy in a large military personnel cohort showed that overweight and obesity had a strong association with the development of the injury. Additionally, this study found that moderate alcohol use was associated with increased risk of Achilles tendinopathy. However, the magnitude of the OR was modest and could be explained by the possible influence of alcohol on metabolic and inflammatory factors . A case-Genetic variations are alterations in DNA sequence among individuals that may account for differences in phenotype and occasionally in health state. Genetic variations occurring in more than 1% of a population are considered useful polymorphisms for genetic linkage analysis. One of the most commonly investigated types of genetic variation is the single-nucleotide polymorphism (SNP), a variation in a single nucleotide located at a specific position in the DNA . The genTNC) and COL5A1 [TNC and COL5A1 genes with the development of Achilles tendinopathy. Tenascin-C regulates cell-matrix interaction in tendon. TNC contained a polymorphism in intron 17 where the number of GT dinucleotide repeats ranged from 3 to 21, and 95% of the alleles contained 12 to 17 GT repeats. Those subjects who were homozygous or heterozygous for the underrepresented alleles of TNC (containing either 13 or 17 GT repeats) were 6.2 times less likely to develop Achilles tendon injury [TNC may be protective from Achilles tendinopathy. COL5A1 encodes the pro-a1 (V) chain of the type V collagen. Two variants of the COL5A1 gene were associated with an increased risk of Achilles tendinopathy. In contrast, a third variant was underrepresented in subjects with Achilles tendinopathy and therefore was associated with a reduced likelihood of Achilles tendinopathy development [COL5A1 gene variants to Achilles tendinopathy in both Australian and South African populations [TNC and COL5A1 gene polymorphisms in genetic predisposition to Achilles tendinopathy.This review includes 14 studies investigating 16 genes linked to collagen structure and tendon homeostasis. Two of these investigated genes located in 9q region, particularly tenascin-C  constructed from one COL27A1 polymorphism (rs946053) and two TNC polymorphisms  showed significant association with Achilles tendinopathy [COL5A1 and MIR608 polymorphisms in South African and Australian cohorts found that a variant of COL5A1 contains a putative polymorphic micro-RNA binding site. The results showed that polymorphisms rs71746744, rs16399 and rs1134170 in COL5A1 and polymorphism rs4919510 in MIR608, which encodes a small micro-RNA, were all independently associated with Achilles tendinopathy, suggesting a role for these four variants on messenger ribonucleic acid (mRNA) stability and the resulting type V collagen synthesis [COL5A1 SNPs rs12722, rs3196378 and rs71746744 was associated with the risk of Achilles tendon pathology [Other studies have assessed the contribution of haplotypes constructed from candidate polymorphisms, in an effort to find interactions between genes or gene products. A collaborative study investigated inopathy . The stuynthesis . A folloynthesis . Howeverathology .MMP3 genotyping had been done on the same cohort as COL5A1 reported by September (2009) [MMP3 rs679620 and COL5A1 rs12722, which are associated with a lower risk of Achilles tendinopathy. Type V collagen is a substrate for MMP3; hence, genetic variants in COL5A1 and MMP3 genes could account for differences in the interactions between the proteins [MMP3 gene’s polymorphisms showed that an inferred haplotype of four SNPs  was associated with Achilles tendinopathy in an Australian cohort [A further study conducted on South African subjects analysed polymorphisms in the gene encoding for matrix metalloproteinase 3 (MMP3) which is involved in the regulation of extracellular matrix homeostasis. Three of the investigated polymorphisms  showed strong association with Achilles tendinopathy, and the most underrepresented haplotype in patients with Achilles tendinopathy indicated that this variant was protective against Achilles tendinopathy . Moreover (2009) , this stproteins . Furthern cohort .COL12A1  and COL14A1  were investigated as they both encode for proteins involved in the biological processes of fibrillogenesis and, like tenascin C, in the modulation of the tendon response to mechanical stress [COL11A1 rs3753841 and rs1676486, COL11A2 rs1799907) have been associated with lumbar disc herniation and rheumatoid arthritis. Whilst none of the polymorphisms were independently associated with Achilles tendinopathy, the construction of a pseudohaplotype consisting of three polymorphisms and the COL5A1 polymorphism rs71746744 revealed a significant association with Achilles tendinopathy. It is hypothesized that the interaction of genes encoding for type V and XI collagens could modulate risk of Achilles tendinopathy, allowing for the possibility that the effects of type XI collagen variants in the developing tendon might affect the structural or functional properties of the mature tendon [Polymorphisms in l stress . Additiol stress . Type XIe tendon .ADAMTS2 rs1054480, ADAMTS5 rs226794, ADAMTS14 rs4747096, ADAMTS12 rs3740199 and TIMP2 rs4789932 gene variants. Researchers found a significant association between the rs4789932 TIMP2 variant and Achilles tendinopathy. The balance between TIMPs and MMPs could be a contributing factor for Achilles tendinopathy development [MMP3 and TIMP2 gene variants in a British cohort showed that a gene variant in TIMP2 rs4789932 was associated with a reduced risk of Achilles tendon pathology in males [FBN2 rs331079 and ELN rs2071307 were studied in Australian and South African cohorts, and the GG genotype in rs331079 was overrepresented in the group with Achilles tendinopathy, indicating an association between fibrillin and injury [TGFB1 and GDF5  were selected as candidates as these proteins had been shown to increase mechanical strength after gene transfection in Achilles tendon in experimentally injured animals [GDF5 rs143383. However, no association with TGFB1 rs1800469 was identified [Genes involved in tendon homeostasis and particularly in the extracellular matrix (ECM) have also been investigated for their association with Achilles tendinopathy risk . The ADAelopment . The repin males . The cond injury . Another animals . This stentified .COL5A1 identified this gene as one of the most likely predisposing factors for Achilles tendinopathy. However, several studies investigating polymorphisms in genes encoding for proteins interacting with type V collagen showed that it is important to be aware of possible connections and pathways whose interactions might be disrupted and therefore alter collagen structure and its functionality and lead to increased or decreased risk of Achilles tendinopathy.A thorough analysis of polymorphisms in IL1RN gene, and its variable number tandem repeat (VNTR) rs2234663 polymorphism has been previously associated with gastrointestinal diseases [IL1β (rs1143627 and rs16944) have been implicated to an increase in IL1β gene expression [COL5A1 gene expression [IL6 (rs1800795) was previously shown to alter IL6 expression [IL1β, IL1RN and IL6, although none of these polymorphisms were associated with the Achilles tendinopathy diagnosis in either of the South African and Australian population groups studied. However, inferred allele combinations constructed from previously studied COL5A1 polymorphisms and these IL1β, IL6 and IL1RN VNTR polymorphisms were associated with an increased risk of Achilles tendinopathy in combined groups [Candidate genes involved in processes surrounding the development of Achilles tendinopathy, such as tendon turnover and inflammation, have also been considered as possible genetic risk factors. The presence of SNPs in cytokine genes, important contributors to the inflammatory response which have been shown to be upregulated in tendinopathy and mechanically loaded tendon were investigated. Interleukin-1β (IL1β) induces inflammatory mediators that upregulate the expression of proteins involved in degradation of the tendon extracellular matrix such as MMPs which target type V collagen. The IL1β receptor antagonist, IL1ra, is encoded by the diseases , osteopodiseases  and athediseases . Whilst pression , interlepression . A SNP ipression  which mad groups . This stCASP) and nitric oxide synthases (NOS) have also been investigated as these molecules had been shown to be involved in pathways accompanying tendon cell apoptosis, and their expression has been found to be elevated in tendinopathy [CASP8 rs3834129, rs1045485, NOS3 rs1799983 and NOS2 rs2779249). A significant association between both CASP8 polymorphisms  and Achilles tendinopathy was found in both populations. The D/D genotype of rs3834129 was associated with tendinopathy, whilst the C allele of rs1045485 was associated with the absence of Achilles tendinopathy. NOS3 (rs1799983) and NOS2 (rs2779249) were not associated with Achilles tendinopathy. However, the data presented in this study showed that the control group in the Australian cohort was not in Hardy-Weinberg equilibrium (HWE), which refers to constant proportions of allele and genotype frequencies in a population, and therefore, this association should be interpreted with caution. Deviations from HWE in the control group may indicate significant methodological flaws including selection bias, population stratification and genotyping errors [Polymorphisms in genes encoding caspases , was investigated as a potential gene associated with Achilles tendinopathy [TNFRSF1A rs4149577 had been associated with several musculoskeletal and inflammatory diseases; however, this study was the first to investigate this polymorphism in association with Achilles tendinopathy. Another polymorphism included in this study was the caspase-3 gene CASP3 rs1049253, which was shown to influence CASP3 mRNA expression. Caspase-3 is involved in cellular apoptosis, including roles in chromatin condensation and DNA fragmentation [CASP8. CNVs are segments of DNA greater than 1 kb in size that can influence phenotypes by changing gene dosage and disruption of coding sequences in DNA. The results did not show any significant association between Achilles tendinopathy, the investigated polymorphisms and CNV. This is possibly due to the study limitations, such as a relatively small sample size and a possible additional degree of error due to the rounding of copy number data into discrete calls, or may indicate that there is no association between these CNVs and Achilles tendinopathy [Tumour necrosis factor receptor 1 gene inopathy . TNFRSF1entation . This stinopathy .While these studies indicate some links between polymorphisms in genes involved in apoptosis and inflammation and Achilles tendinopathy, the majority of these studies were unable to demonstrate that these genes are probable risk factors for Achilles tendinopathy. The processes of apoptosis and inflammation clearly play a role in the pathology of Achilles tendinopathy; however, further investigation should be undertaken to clarify the role of genetic contributors in this pathological processes.It is clear from previous research that biomechanical issues and training load errors are risk factors for Achilles tendinopathy . This sy


Well-developed coronary collateral circulation usually results in fewer infarct size, improved cardiac function, and fewer mortality. Traditional coronary risk factors  have some effects on coronary collateral circulation. However, the association between these risk factors and coronary collateral circulation are controversial. Given the confusing evidences regarding traditional cardiovascular risk factors on coronary collateral circulation, we performed this meta-analysis protocol to investigate the relationship between traditional risk factors of coronary artery disease and coronary collateral circulation.MEDINE, EMBASE, and Science Citation Index will be searched to identify relevant studies. The primary outcomes of this meta-analysis are well-developed coronary collateral circulation. Meta-analysis was performed to calculate the odds ratio (OR) and 95% confidence interval (CI) of traditional coronary risk factors . Pooled ORs were computed as the Mantel–Haenszel-weighted average of the ORs for all included studies. Sensitivity analysis, quality assessment, publication bias analysis, and the Grading of Recommendations Assessment, Development and Evaluation approach (GRADE) will be performed to ensure the reliability of our results.This study will provide a high-quality synthesis of current evidence of traditional risk factors on collateral circulation.This conclusion of our systematic review and meta-analysis will provide evidence to judge whether traditional risk factors affects coronary collateral circulation.Ethics and dissemination: Ethical approval is not required because our systematic review and meta-analysis will be based on published data without interventions on patients. The findings of this study will be published in a peer-reviewed journal. Although this so-called smokers’ paradox has been attributed to the younger age, lower co-morbidity, more aggressive treatment and lower risk profile of the smoker, well-developed coronary collateral circulation may be another explanation. Some investigators have found that smoking was associated with well-developed coronary collateral circulation which might alleviate myocardial ischemia and necrosis when epicardial coronary arteries were occluded.,4 Well-developed coronary collateral circulation usually results in fewer infarct size, improved cardiac function, and fewer mortality.–7 Smoking improves coronary collateral circulation which may partial explain the so-called smokers’ paradox. But some researchers found opposite evidence that smoking caused rarefaction of coronary collateral circulation.An increasing number of people in developing countries are under traditional cardiovascular risk factors such as smoking, hypertension and diabetes mellitus, et al, which may result in high rates of cardiovascular disease. Yetkin et al found that diabetes mellitus was an independent factor for rarefaction of coronary collateral circulation. However, the study by Niccoli et al found that patients with diabetes mellitus had better collateral development compared with nondiabetic patients. Although, the association between these risk factors and coronary collateral circulation is controversial, these risk factors do have some effects on collateral on coronary collateral circulation. Given the confusing evidences between traditional cardiovascular risk factors and coronary collateral circulation, we performed this meta-analysis protocol to investigate the relationship between traditional risk factors of coronary artery disease and coronary collateral circulation.Except from smoking, other cardiovascular risk factors such as hypertension, diabetes mellites also have some impacts on coronary collateral circulation.2 We conducted a review of traditional coronary artery disease risk factors and coronary collateral circulation by systematically searching for relevant studies in which traditional risk factors  was all or part of the exposure of interest or where traditional risk factors had been part of a subgroup analysis.This systematic review and meta-analysis protocol has been registered with the International Prospective Register of Systematic Reviews (CRD42018087821). This protocol is reported following the Preferred Reporting Items for Systematic Review and Meta-Analysis Protocols (PRISMA) guidance.33.1Any case–control study was included in our meta-analysis if the extent of coronary collateral circulation of each patient could be extracted according to different traditional risk factors .3.2 But we will include those studies did not define CTO exactly. Studies that included patients with severe stenosis  were excluded.Studies that had patients with chronic total occlusion (CTO) of the main branches of the coronary arteries were included. Ideally, CTO in the included studies should be defined as a totally occluded segment with Thrombolysis in Myocardial Infarction flow grade 0 and an estimated duration of at least 3 months.3.3Traditional coronary risk factors such as smoking, diabetes, and hypertension were defined as exposure factors. Patients were divided into 2 groups according to patients with or without traditional risk factors .3.4 Well-developed coronary collateral circulation was defined as Rentrop score 2 or 3.The primary outcomes of this meta-analysis are well-developed coronary collateral circulation. The definition of collateral circulation is Rentrop scoring system: 0= no collateral vessels; 1= thread-like, poorly opacified collaterals without visualization of the epicardial artery; 2 = partial filling of the epicardial artery via collateral channels; 3 = complete filling of the epicardial segment of the artery via collateral channels.3.5The databases will be searched to obtain relevant studies included MEDINE, EMBASE, Science Citation Index before January 1, 2018. Language is restricted to English. We will use keywords, MeSH term searches and Emtree to find relevant studies. We also will search the reference lists of relevant studies and reviews, editorials, and letters, together with related conference abstracts to identify further articles.44.1Relevant studies were searched by 2 independent investigators (JYP and ZHX). We use Endnote  to manage relevant articles and remove duplicated articles. Disagreements were resolved by consensus or a third investigator (XQH). The process of study search strategy will be shown in a PRISMA-compliant flow chart Fig. .4.2We abstracted the following data from the selected articles: study characteristics ; characteristics of included participants ; definition of coronary collateral circulation; and risk factors of coronary collateral circulation. All these data were extracted to a prepiloted, standardized excel sheet. When relevant data were missing, study authors were contacted by e-mail, before excluding the reference for inaccessibility of data.4.3 A maximum of 2 stars can be given for comparability. A star system of the NOS  has been performed for the evaluation  for quality of case–control and cohort studies.on Table .4.4I2 statistic, and a P-value ≤.05 is considered statistical significant.Meta-analysis will be performed to calculate the odds ratio (OR) and 95% confidence interval (CI) of traditional coronary risk factors . Pooled ORs are computed as the Mantel–Haenszel-weighted average of the ORs for all included studies. Since the true treatment effect of various postconditioning protocols may have varied among the included trials, the random-effects model is used in the analysis. Statistical heterogeneity among the trial-specific ORs is checked and quantified by the If quantitative analysis is not appropriate, we will just perform a narrative, qualitative summary and the results will be presented with text and tables.4.5Sensitivity analysis will be performed to assess the contribution of each study to the pooled estimation by excluding one trial at a time and recalculating the pooled ORs estimation for the remaining studies. Both random efforts and fixed-effort model will be used both.4.6,16 If publication bias does exist, we use “trim and fill” method to analyze publication bias on our outcomes.In order to study publication bias on our meta-analysis, publication bias will be qualitatively analyzed by funnel plots as well as qualitatively analyzed by Egger's tests.4.7 This approach will assess the risk of bias, directness of evidence, precision of estimate, heterogeneity, publication bias, effect size, and plausible explanation of the confounding or bias. The quality of evidence will be listed as high, moderate, low, or very low. We will perform this analysis using GRADEpro online software.The overall evidence then assessed by the Grading of Recommendations, Assessment, Development, and Evaluation (GRADE) approach.5Currently, the association between traditional coronary artery disease risk factors and coronary collateral circulation is uncertain. Contradictory evidences and opinions continue to surface. Therefore, it is necessary to perform a high quality systemic review and meta-analysis, in which our rigorous approach will provide a solid evidence for these issues. To our knowledge, this will be the first systematic review and meta-analysis using data of case–control studies to ensure the association between traditional risk factors and collateral circulation for patients with CTO.Conceptualization: Zhenhua Xing, Xinqun Hu.Data curation: Zhenhua Xing.Formal analysis: Zhenhua Xing.Writing – original draft: Junyu Pei.Writing – review & editing: Liang Tang.


P = 0.016; I2 = 51.5%). No statistical differences were detected in women. Subgroup analysis of different racial groups revealed significant correlation between the SCARB1 rs5888 polymorphism and higher high-density lipoprotein cholesterol (HDL-C) levels  and lower triglyceride levels  in non-Asian men. No evidence of heterogeneity was observed when eliminating outlier studies, and no publication bias was detected. This meta-analysis suggests the SCARB1 rs5888 polymorphism is associated with higher HDL-C and lower triglyceride levels in non-Asian men.Studies have shown that the scavenger receptor class B type 1 (SCARB1) rs5888 polymorphism impacts fasting blood lipid levels differently in men and women. A meta-analysis and statistical tests was therefore performed to determine the relationship between the rs5888 polymorphism and lipid levels in men and women. Twelve studies with 12,147 subjects were selected for this study. In a dominant model, the CT+TT genotype group had lower triglyceride levels than the CC group in men (standardized mean difference (SMD): −0.11; 95% confidence interval (CI): −0.21 to −0.02;  Cardiovascular disease (CVD) is the leading cause of death in developed and many developing countries . Risk faSCARB1 is an HDL receptor closely coupled with HDL-C that promotes HDL-C uptake and reverse translation . StudiesWe searched for cross-sectional and case-control studies from January 1990 to June 2015 in PubMed, Embase, China National Knowledge Infrastructure, Wanfang Data, and China Biology Medicine databases. The keywords used for this search were as follows: “scavenger receptor AND (genetic OR variant OR variation OR mutation OR mutated OR polymorphism OR snp) AND [(low-density lipoprotein cholesterol) OR (high-density lipoprotein cholesterol) OR LDL-C OR HDL-C or triglyceride or ].” The reference lists of the included studies were also searched for relevant results.Two investigators (Li-fang Ye and Ya-ru Zheng) independently reviewed all the studies retrieved from the database searches. We followed guidelines from the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement Figure . The selTwo reviewers (Li-fang Ye and Ya-ru Zheng) extracted the papers independently. We initially reviewed the article titles and abstracts and then excluded those that did not fit the inclusion criteria. All the included studies were extensively reviewed. Any disagreements were resolved by discussion. For each included study, we assessed methodologic rigor and quality using STREGA (Strengthening the Reporting of Genetic Association Studies) guidelines, an extension of STROBE  guidelines specifically developed to assess quality of genetic association studies. The information that was extracted included first author name, year of publication, ethnicity, sample size, sex, age, health condition, genotype information , and relationship between genotypes and fasting blood lipid levels.All data were expressed as mean ± standard deviation. All analyses were performed using the STATA software package v12.0 . We used a dominant model (CT + TT vs CC) as a result of the low rate of the TT genotype in the Asian population. When one article included more than one subpopulation , the subpopulation was analyzed as a separate comparison. In addition, we divided the ethnic subgroups into Asian and non-Asian. The meta-analyses on the subgroups were performed with at least four comparisons to ensure adequate statistical power.2-based Q-tests. I26 values of 25%, 50%, and 75% represent low, medium, and high levels of heterogeneity, respectively. Potential heterogeneity was detected using the Galbraith plot, and the pooled SMD was recalculated with the outlier studies removed. Publication bias was assessed using the Begg's funnel plot and the Egger's linear regression test. P values less than 0.05 were considered statistically significant.The differences in the variables between the genotypes were expressed as the pooled standardized mean difference (SMD) with 95% confidence interval (CI), and the random effects model was used for all analyses . We asseInitial search of the literature yielded 581 publications. Forty-three studies had subject overlap with other publications. Sixty-three publications were reviews. Four hundred three studies were excluded based on their title and abstract. Next, full-text articles were retrieved and assessed based on inclusion criteria. Overall, 60 articles were excluded for the following reasons: 44 articles did not provide complete data for this meta-analysis, and 16 articles presented data on other polymorphisms. Finally, 12 studies , 24 wereThe characteristics of the 12 included studies are summarized in Table P = 0.016; I2 = 51.5%) than the CC genotype group      in the non-Asian population  Figure , LDL-C  Figure , and TC ) Figure . Subsequ) Figure  and lowen Figure . No signP = 0.248; I2 = 0%)     Figure , LDL-C  Figure , TC (SMD) Figure , and TG ) Figure . The eth2 values of heterogeneity were less than 50%. Only the levels of heterogeneity of TG in men and LDL-C and TC in women were medium. To identify the studies that contributed to this heterogeneity, a Galbraith plot analysis was performed. The results indicated that a single study [P = 0.035, I2 = 4.9%, Pheterogeneity = 0.396; LDL-C in women: SMD: 0.01, 95% CI: −0.08 to 0.10, P = 0.868, I2 = 20%, Pheterogeneity = 0.283; TC in women: SMD: −0.03, 95% CI: −0.10 to 0.03, P = 0.326, I2 = 0%, Pheterogeneity = 0.494).The majority of results for the fasting blood lipid levels revealed that the Ile study  was the le study , 17, 20 P = 0.427; women: t = 0.37, P = 0.715), LDL-C , TC , and TG . The shapes of the funnel plots do not show evidence of asymmetry (The Egger's test revealed that there was no publication bias in the analyses for HDL-C (men: t = −0.83, symmetry .Various candidate genes have been reported as predisposing factors for dyslipidemia, including those involved in lipid transport and metabolism . It is wTo our knowledge, we present here the first meta-analysis to investigate the association between SCARB1 polymorphisms, sex, and lipid levels. We included a total of 12 publications including 12,147 subjects. Our results demonstrated a significant correlation between the SCARB1 rs5888 polymorphism and TG levels that was dependent on the presence of the T allele in men. No significant association between rs5888 polymorphism and plasma lipid levels was detected in women. The ethnicity subgroup analyses indicated that the rs5888 polymorphism was significantly correlated with higher HDL-C levels and lower TG levels in non-Asian men but not in Asian men. No significant association between SCARB1 rs5888 polymorphism and plasma lipid levels was identified in either Asian or non-Asian women.SCARB1 has been identified as a physiologically important HDL receptor; however, it is also a multiligand receptor participating in the metabolism of other plasma lipoproteins . The mec2 values were less than 50%. Medium levels of heterogeneity were observed for TG levels in men and LDL-C and TC levels in women. This heterogeneity could be attributed to the variable genetic and environmental factors influencing the different groups in the included studies. Fasting blood lipid levels and CVD are affected by both genetic factors and environmental factors, such as diet, lifestyle, and physical activity. The heterogeneity was effectively removed or decreased, but the SMD values and their 95% CIs did not change significantly after exclusion of the outlier studies by using a Galbraith plot. This supported the reliability of the results in our meta-analysis.High heterogeneity can potentially affect the interpretation of results. In our study, heterogeneity was relatively low, considering that most of the IHowever, a few limitations of this meta-analysis should be considered. First, the sample size of the included studies was relatively small and might have been insufficient to achieve adequate statistical power for detecting any small effects. A larger sample size is required to further investigate any association between the SCARB1 polymorphism and blood lipid levels. Second, dyslipidemia is also influenced by other genes and environmental factors; however, this was not addressed in this study due to the lack of data from the included studies. A more comprehensive investigation of the interactions of the rs5888 polymorphism with other polymorphic loci or environmental factors and lipid levels is needed.In conclusion, our analysis demonstrated that the SCARB1 rs5888 polymorphism is associated with higher HDL-C levels and lower TG levels in non-Asian men. Additional studies with larger sample sizes should be conducted to clarify the association between SCARB1 polymorphisms, sex, and lipid levels. Studies investigating the functions of these polymorphisms are also needed.


However, no association between miR-17-92 family expression and cancer progress-free survival (PFS) was found (p > 0.05). Subgroup analyses showed that high expression of miR-17-92 family was associated with poor OS  and DFS  among the Asian, and no association was found for the Caucasian (p > 0.05). Besides, the HRs of miR-17-92 family high expression in tissue and serum samples was 1.68 (1.35–2.09) and 2.20 (1.08–4.46) for OS, and 1.73 (0.80–3.74) and 3.37 (2.25–5.02) for DFS. It also found that high expression of miR-17-92 family predicted a poor OS in breast cancer, esophageal squamous cell carcinoma, lymphoma and other cancers. Findings suggest that miR-17-92 family can be an effective predictor for prognosis prediction in cancer patients.Recent evidence indicates that miR-17–92 family might be an essential prognostic biomarker for human cancers. However, results are still inconsistent. We therefore performed a meta-analysis to evaluate the predictive role of miR-17–92 family in human cancer prognosis. We searched literatures published before March 31th, 2017 inPubMed, Cochrane and Embase databases. Twenty six studies were included in our analyses. The overall hazard ratios (HRs) showed that high expression level of miR-17-92 family was a predictor of poor overall survival (OS): adjusted HRs = 1.71, 95% confidence intervals (CIs): 1.39–2.11,  Cancer is now the second leading cause of death worldwide after cardiovascular disease, with cases of nearly all types of cancer on the rise . GLOBOCAMiRNAs are small non-coding RNAs of approximately 20–22 nucleotides in length that play vital roles in the regulation of gene expression at the post-transcriptional level . AlteratThe miR-17-92 family maps to human chromosome 13 (13q31.3) and encodes for the miR-17–92 cluster  and two paralogs  . SeveralContradictory results in previous studies may due to small sample sizes and different detection methods. Further studies are needed to explore the relationship between miR-17-92 family expression and the prognosis of cancer patients. Evidence from previous studies indicate that clustered miRNAs with similar sequences may regulate a set of mRNA targets and therefore function as powerful regulators of specific cellular activities . A clustA total of 1072 articles were retrieved from the PubMed, Cochrane, and Embase databases, of which 969 were excluded for irrelevant to cancers; 40 were duplicates; 21 were abstracts, review articles or letters; 10 articles did not contain available data; 2 studies used cell lines; and 1 study used inappropriate controls (did not use tumor tissues or serum samples). Finally, a total of 29 studies included in these analyses were assessed for quality using the Newcastle–Ottawa scale (NOS). Three studies with low quality  were then excluded. Eventually, 26 , 20–42 sThe main characteristics of included studies included in the present analysis was summarized in Table A total of 23 studies were included for OS analyses, of which 15 , 40, 42 I2 = 88%, p < 0.00001). In random effect model, results showed that higher expression level of miR-17-92 family was associated with poor OS  (p = 0.0004), while no association was found for the Caucasian (p = 0.0005) and serum samples  (p < 0.05)  . Subgrouaucasian . Subgrou0.00001) . Results < 0.05) . (Table p < 0.00001). A moderate between-study heterogeneity was found  . However, no association was found for the Caucasian  (p < 0.00001) and serum samples  (p = 0.002), ESCC , BL  and other cancers . No associations was found in CRC, lung cancer, HCC, gliomas and PC (p > 0.05)  . Subgrou = 0.03) . In subg > 0.05)  , and thus, the random effect model was applied to calculate the pooled HRs and its 95% CIs. Results showed that no association between high expression of miR-17-92 family and cancer DFS   . (Table  = 0.41) .p = 0.0008), and a high between-study heterogeneity was found  , while no associations was found in the Caucasian (p > 0.05) (p < 0.00001)  . In subg0.00001)  . The funnel plot also showed a symmetrical distribution, suggesting no publication bias in the overall analysis of included studies . However, no association was found after the adjusted value was calculated   Figure . (Table Only three studies were included to determine the association between the expression of miR-17-92 family and cancer PFS. Thus the publication bias could not well reflect in the funnel plot and Begg's test.p < 0.05), while no association was found for cancer DFS (p > 0.05). In adjusted analyses, high expression of miR-17-92 family was found associated with poor cancer OS and DFS (p < 0.05), but no association was found for cancer PFS(p > 0.05). Results indicated that increased miR-17-92 family expression plays an important role in cancer OS, and the miR-17-92 family may be a promising biomarker to predict OS in cancer patients. Inconsistent results in PFS and DFS may cause by several reasons. Firstly, pooled cude HRs was calculated from data of univariate analyses of included studies, which did not eliminated effects of confounding factors. The pooled adjusted HRs was computed according to the data of multivariate analyses. Adjusted HRs considered effects of confounders and were more representative than the pooled crude HRs. Secondly, studies included in unadjusted and adjusted analyses were not identical. Some studies reported results of univariate and multivariate analyses simultaneously; while others only presented results of the univariate or multivariate analyses. Thus, we calculated the pooled HRs value separately according to analytical methods used in included studies. Thirdly, confounding factors varied in different studies, which may lead to inconsistency in results. In addition, only 3 studies were included in pooled PFS analysis. Theoretically, the merged data could be affected easily by studies with large sample size. Although some inconsistent results were found in our analysis, the miR-17-92 family still played an important role in cancer prognosis. Inconsistent pooled results implied that the role of miR-17-92 family in cancer prognosis may be complex and controversial.Results of the unadjusted analyses showed that higher expression levels of miR-17-92 family were associated with poor OS and PFS (p = 0.034) was significantly associated with poor survival in patients with human head and neck squamous cell carcinoma (HNSCC). Besides, Zakrzewska et al. ; (2) Comparability: including one item [Item 5: Comparability of cohorts on the basis of the design or analysis]; and (3) Outcome: including three items . Total scores of NOS ranged from 0 to 9; studies with NOS scores ≥ 6 were considered as high-quality studies, which studies scores below 6 were regarded low-quality. Any discrepancies were discussed with the third investigator (XYL).p values were used to estimate the effect across studies for the association between miR-17-92 family and cancer prognosis. Prognosis outcomes mainly contained the OS, DFS, and PFS. Heterogeneity among studies was evaluated using I2 statistics. The values of I2 less than 25, 50, and 75% indicated low, moderate, and high degree of heterogeneity, respectively. When significant heterogeneity was observed (I2 > 50%), pooled HRs estimations were calculated using a random effect model. Otherwise, a fixed effect model was applied [Z-test, and a p value less than 0.05 was considered statistically significant [p value less than 0.05 was considered to indicate significant publication bias. All analyses were conducted using RevMan (Version 5.2) software and Stata 12.0.Pooled HRs, 95% CIs, and  applied . Sensiti applied . Subgrounificant . Publicanificant . A p val