A number of studies have demonstrated the critical role of long non-coding RNA gastric cancer high expressed transcript 1 (GHET1) in many cancers. This meta-analysis provides an evidence-based evaluation of the prognostic role of GHET1 in cancer.Literature searches were conducted in several databases including Medline, Cochrane, EMBASE, CNKI, and Wanfang. The pooled odds ratio (OR) and hazard ratio (HR) with 95% confidence interval (CI) were used to evaluate the role of GHET1 in cancer. The study protocol was registered at PROSPERO (ID: CRD42018111252).Sixteen studies, containing 1315 patients, were analyzed in this meta-analysis. The pooled results indicated that GHET1 overexpression was significantly associated with poor overall survival (OS) and disease-free survival (DFS) in cancer. Moreover, up-regulation of GHET1 expression predicted larger tumor size, positive lymph node metastasis, positive distant metastasis, and advanced TNM (tumor-node-metastases) stage in human cancers.There is a significant correlation between up-regulation of GHET1 and both poor prognosis and advanced clinicopathological cancer characteristics. GHET1 may be a potential prognostic predictor for human cancers. Despite great progress in cancer diagnosis and treatment, there were still 18.1 million new cancer cases and 9.6 million cancer deaths worldwide in 2018 –5.Recently, many studies have indicated that long non-coding RNAs (lncRNAs) play a crucial role in the progression of cancer –10. SomeClinically, several studies have indicated that up-regulation of GHET1 is associated with poor prognosis and advanced clinical features in several cancers –24. MostThis systematic meta-analysis was conducted accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines. It has been registered with PROSPERO (ID: CRD42018111252).Literature searches were conducted in several databases, including Medline, Cochrane, EMBASE, CNKI, and Wanfang, up until April 15th, 2019. The search strategy was as follows: “GHET1” OR “lncRNA GHET1“ OR “gastric cancer high expressed transcript 1“) AND (“cancer“ OR “neoplasm“ OR “tumor“ OR “carcinoma“). The search was limited to English and Chinese studies. The references of relevant studies were also retrieved to avoid missing any potentially eligible studies.The inclusion criteria for this meta-analysis were as follows: (1) detection of GHET1 expression in human cancers by quantitative real-time PCR (qRT-PCR); (2) patients in the study were divided into subgroups based on different GHET1 expression levels; (3) prognosis or clinicopathological feature of GHET1 was reported; (4) hazard ratios (HRs) and 95% confidence intervals (CI) were able to be obtained directly or indirectly from the article. In addition, the exclusion criteria were as follows: (1) reviews, editorials, conference reports, case reports, and meta-analyses; (2) non-human tissue studies; (3) studies only investigating the molecular mechanisms of GHET1; (4) duplicate publications. The titles and abstracts were first evaluated based on the inclusion and exclusion criteria; the full texts of those reports that appeared to meet the criteria were then further evaluated.Two researchers (Dingding Wang and Xiaolian Fang) independently extracted data from the selected studies according to uniform data extraction standards; any disagreements were settled by consensus with a third investigator (Hong Zhang). Extracted data included: first author’s name, publication year, country or region, sample size, cancer type, method for detection of GHET1, cut-off values, treatment data, disease-free survival (DFS), overall survival (OS), and clinical stage of cancer. If the HRs and 95% CIs for DFS or OS were not available in the paper, the data were indirectly extracted from survival curves, based on the approach described previously . The Newhttp://cancergenome.nih.gov/publications/publicationguidelines). GEPIA was used to analyze RNAseq data. Differential expression analysis was conducted using one-way ANOVA. Survival analysis was performed using the Kaplan–Meier method and log-rank test.TCGA data, including RNAseqV2 and clinical data, were extracted from the TCGA Data Portal and UCSC Xena project, according to the publication guidelines . Moreover, Engauge Digitizer 10.0 was utilized to extract HRs and 95% CIs from the Kaplan–Meier survival curves. A P value of less than 0.05 was considered to be statistically significant.Pooled HRs and 95% CIs were used to assess the relationship between GHET1 expression and prognosis. Odd ratios and 95% CIs were combined to evaluate the relationships between GHET1 expression and clinicopathological factors. If the 95% CI of the combined OR or HR did not overlap 1, the result was considered statistically significant. The IA preliminary literature search was conducted based on the established search strategy; a total of 107 articles were obtained. After removing 75 duplicates, the titles and abstracts of the remaining 32 articles were evaluated according to the inclusion criteria. Following this, six studies were excluded. The full texts of the remaining 26 articles were reviewed; this resulted in the exclusion of 10 papers due to a lack of eligible prognostic or clinicopathological information. Finally, a total of 16 studies were included in the meta-analysis . The pooled HR for the high GHET1 expression group versus the low group was 2.037 . This pooled result indicates a significant association between overexpressed GHET1 and poor OS.Twelve studies reported OS for eight types of cancer based on GHET1 expression in 1114 patients. As shown in Fig. In order to further explore the potential prognostic value of GHET1, a series of subgroup analyses were performed based on cancer type, sample size, cut-off value, treatment, and NOS score. The results indicated that, regardless of the cancer type, sample size, cut-off value, treatment, and NOS score, the up-regulation of GHET1 was significantly correlated with poor OS in all subgroup analyses Table .Table 2S2 = 37.1%, P = 0.190). The pooled results indicated that elevated GHET1 expression was significantly correlated with shorter DFS , kidney renal papillary cell carcinoma (KIRP), liver hepatocellular carcinoma (LIHC), stomach adenocarcinoma (STAD), colon adenocarcinoma (COAD), breast invasive carcinoma (BRCA), bladder urothelial carcinoma (BLCA), and esophageal carcinoma (ESCA) (|Log2FC| Cutoff > 1, q-value < 0.01, Fig. Accumulating evidence has demonstrated that lncRNAs act as crucial regulators of almost all aspects of physiological and pathological processes –30. MultIn the present study, we evaluated the prognostic value of GHET1 in cancer. The pooled HR indicated that GHET1 overexpression was significantly associated with poor OS and DFS in cancer. Moreover, further subgroup analyses indicated that elevated GHET1 expression was significantly correlated with OS in each subgroup, regardless of the analysis model, sample size, cut-off value, treatment, cancer type, and NOS score. The pooled data illustrated that GHET1 overexpression was significantly associated with larger tumor size, positive lymph node metastasis, positive distant metastasis, and advanced TNM stage. Unexpectedly, we failed to identify an association between GHET1 expression and histological differentiation. However, these results might be not reliable because there was significant heterogeneity among the included studies. Overall, high GHET1 expression was an unfavorable risk factor for survival outcomes in patients with cancer; thus, GHET1 might be a valuable biomarker for a variety of cancers. To our knowledge, this research is the first meta-analysis focusing on the prognostic value of GHET1 in human cancers.Many studies have tried to illustrate the correlation between high GHET1 expression and cancer prognosis; however, the molecular mechanism of GHET1 remained unclear , 23, 24.The present meta-analysis has several limitations. First, this meta-analysis included only 16 studies, and all of these studies were from China. Therefore, the results might only apply to Asian or Chinese patients, which may limit the representativeness of the results. The validation tests using data from TCGA make up this disadvantage in some extent. Second, the HRs and 95% CIs in several studies could not be directly obtained. Thus, we extracted the data from the Kaplan–Meier curve in these studies, which might introduce statistical errors. Finally, the sample sizes of some cancer types in this meta-analysis were limited; this may have contributed to the heterogeneity and may have affected the reliability of the pooled results for some cancer types. Nevertheless, the results of this meta-analysis should be verified by studies evaluating more cancer types with larger sample sizes.In sum up, the up-regulation of lncRNA GHET1 expression was significantly associated with poor OS, poor DFS, and advanced clinicopathological characteristics in various cancers. GHET1 can be considered to be a promising prognostic predictor for human cancers. However, high-quality studies with larger samples sizes and those encompassing more cancer types are still needed to verify these conclusions.Additional file 1: Fig. S1. OS plots for each TCGA cohort.


To assess the association of sleep duration and quality with the risk of preterm birth.Relevant studies were retrieved from the PubMed and Web of Science databases up to September 30, 2018. The reference lists of the retrieved articles were reviewed. Random effects models were applied to estimate summarized relative risks (RRs) and 95% confidence intervals (CIs).I2 = 57.4%). Additionally, as compared with women with good sleep quality, the summary RR was 1.54 (95% CI = 1.18–2.01) for women with poor sleep quality , with high between-study heterogeneity (I2 = 76.7%). Funnel plots as well as the Egger’s and Begg’s tests revealed no evidence of publication bias.Ten identified studies (nine cohort studies and one case-controlled study) examined the associations of sleep duration and quality with the risk of preterm birth. As compared with women with the longest sleep duration, the summary RR was 1.23 (95% CI = 1.01–1.50) for women with the shortest sleep duration, with moderate between-study heterogeneity (This systematic review and meta-analysis revealed that short sleep duration and poor sleep quality may be associated with an increased risk of preterm birth. Further subgroup analyses are warranted to test the robustness of these findings as well as to identify potential sources of heterogeneity. Each year, more than 10% of births worldwide are preterm, occurring before gestational week 37. Prematurity, which is a leading cause of mortality among children aged less than 5 years , leads tDue to psychophysiological changes caused by pregnancy, approximately 27.9% of women sleep for less than 7 h per night because of sleep disturbances . Thus, nA previous meta-analysis evaluated the association between sleep quality and duration with the risk of preterm birth , but theAfter formulating the study questions, a systematic review and meta-analysis were performed in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines . Based oThe titles and abstracts of the retrieved articles were screened by two independent reviewers (LW and FJ) and any disagreements were resolved by discussion. Articles that fulfilled the following criteria were eligible for the present meta-analysis: (i) participants were pregnant women who were recruited from all populations; (ii) sleep duration and quality were quantified during the term of pregnancy; (iii) preterm birth, as defined as childbirth before gestational week 37, was the primary or secondary outcome; (iv) the study used an observational design ; and (v) the relative risk (RR), hazard ratio (HR), or odds ratio (OR), and corresponding 95% confidence interval (95% CI) or necessary data for calculation were reported. Review articles, case reports, commentaries, and conference abstracts were excluded. If data were duplicated in more than one study, the study with the largest number of cases was included.The data from each of the included studies were extracted carefully by two independent reviewers (LW and FJ). The following information from each article was recorded in a standardized, pre-designed spreadsheet: the first author’s name; publication year; country of origin; study design; number of cases/participants; source of patient recruitment; sleep assessment; pregnancy outcome categories; exposure categories (period of exposure measurement); risk estimates from the most fully adjusted model with associated 95% CI values; and confounders adjusted for multivariate analysis.Considering that all of the included articles were observational studies, two independent reviewers (LW and FJ) used the Newcastle–Ottawa quality assessment scale for observational studies to assess the risk of bias . SubsequFor studies that separately reported results for different trimesters, but not in combination, inverse variance-weighted fixed effects meta-analysis was first used to generate a risk estimate of the overall study at full gestation before random-effects meta-analysis. For studies that failed to use the category with the shortest sleep duration as the reference, the effective count method proposed by Hamling et al.  was usedI2 statistic. I2 values of 25, 50, and 75% were considered to represent low, moderate, and high heterogeneity, respectively [I2 was < 50%. The RR estimates were then assessed for consistency [p < 0.10). The G*Power 3.1.9.2 statistical power analysis program was used to calculate the sample size at an alpha value of 0.05, power of 80% (β = 0.2), estimated RR of 1.2, and proportion of outcome in the exposure group of 0.3. Accordingly, the lowest sample size for a proper cohort was n = 1138 for dichotomous outcomes. All data analyses were performed using STATA software, version 12.0 . A probability (p) value of < 0.05 was considered statistically significant unless otherwise specified.Between-study heterogeneity of each meta-analysis was estimated using the ectively . Summaryectively  was usedsistency . PotentiOf the 3266 retrieved studies remaining after the removal of duplicates, 3233 (99.0%) were excluded after screening of the titles and abstracts. After full text review of the remaining 33 1.01%) studies, 16 analyzed diseases and symptoms related to sleep disorders, and seven lacked OR and 95% CI values. Among these seven studies, three suggested no association between sleep and risk of preterm birth –26, two .01% studn = 5), India (n = 1), Greece (n = 1), Japan (n = 1), and China (n = 1), and were adjusted for potentially important confounders, such as body mass index (n = 4) and maternal age (n = 5). On the basis of the Newcastle–Ottawa quality assessment scale, all studies were judged to have a low risk of bias (Table The characteristics of the 10 studies (nine cohort studies and one case-control study) included for analysis are summarized in Tables ndex n =  and mateI2 = 57.4%, p = 0.029). Funnel plots as well as the Egger’s and Begg’s tests  revealed no evidence of publication bias .The summarized RR values ranged from 1.41  to 1.40i et al. ). After I2 = 76.7%, p < 0.001). Funnel plots as well as the Egger’s and Begg’s tests  revealed no evidence of publication bias  were similar to the original estimates. Additionally, the summarized RR values ranged from 1.26  to 2.05n et al. ).The present systematic review and meta-analysis of 10 observational studies demonstrated that short sleep duration and poor sleep quality were associated with a significantly increased risk of preterm birth. This association highlights the vital significance of pregnant women to reduce the risk of premature birth.n = 1091), in which 23% of pregnant women reported a sleep duration of ≤5 h in the third trimester. However, Reutrakul et al. [n = 116). Meanwhile, Li et al. [n = 48) of the sample [In regard to sleep duration and quality, the inconsistent findings of previous studies might be attributed to differences in the trimester examined and geographical location. For instance, Micheli et al.  conductel et al.  reportedi et al.  enrolledi et al.  reportedi et al. , 38. Wari et al.  speculati et al. , 36, whii et al. . In addie sample .The findings of the current study raise questions about the potential mechanisms underlying the increased risk of preterm birth due to sleep disorders. Sleep deprivation partially accounts for the proinflammatory cytokine response –41, immun = 5693) and instances of preterm birth associated with sleep duration (n = 1248) and sleep quality (n = 156). The large sample size of this meta-analysis provide strong power for the main analyses and the conclusions derived. Furthermore, numerous sensitivity analyses showed that the main findings were robust. Of note, quality assessment showed that all of the included studies were at a low risk of bias.The strengths of the present meta-analysis lie in its quantitative analysis of the association of sleep duration and quality with the risk of prematurity using a large number of participants (Findings from the present meta-analysis should be interpreted in light of several limitations. First, the present meta-analysis was prone to inherent recall and selection bias due to the inclusion of original observational studies. Although case-control studies are more susceptible to bias than cohort studies, the results were robust after exclusion of the only case-control study from the analyses. Furthermore, the PSQI is an important clinical and research tool to gauge sleep quality . HoweverSome common chronic diseases, as mediators between short sleep duration and preterm birth, such as diabetes , 52, hypIn summary, the results of this meta-analysis provide vital insights into the association of short sleep duration and poor sleep quality with a significantly increased risk of preterm birth. These findings may help researchers to identify women at risk of pre-term birth early during pregnancy to provide targeted interventions.


Anaerostipes genus was associated with higher estimated glomerular filtration rate (eGFR) in the overall population  and non-diabetes mellitus (DM) subgroup , while it had a non-significant association with the risk of CKD and eGFR in DM patients. Subjects with higher abundance of Desulfovibrio spp. had a significantly lower level of eGFR ; the same results were observed in non-DM  subjects. Acidaminococcus, Bacteroides, Bifidobacterium, Faecalibacterium, Lactobacillus and Megamonas had no significant association with eGFR in the overall population, DM and non-DM subgroups ; they also presented no significant association with the risk of CKD . Analyses of MR-PRESSO did not highlight any outlier. The pleiotropy test, with very negligible intercept and insignificant p-value, also indicated no chance of pleiotropy for all estimations. The leave-one-out method demonstrated that the observed links were not driven by single single-nucleotide polymorphism. Conclusions: Our results suggest an adverse association of Desulfovibrio spp. and a beneficial association of Anaerostipes spp. with eGFR. Further studies using multiple robust instruments are needed to confirm these results.Background: The microbiota composition is now considered as one of the main modifiable risk factors for health. No controlled study has been performed on the association between microbiota composition and renal function. We applied Mendelian randomization (MR) to estimate the casual impact of eight microbiota genera on renal function and the risk of chronic kidney disease (CKD). Methods: MR was implemented by using summary-level data from the largest-ever genome-wide association studies (GWAS) conducted on microbiota genera, CKD and renal function parameters. The inverse-variance weighted method (IVW), weighted median (WM)-based method, MR-Egger, MR-Robust Adjusted Profile Score (RAPS), MR-Pleiotropy RESidual Sum and Outlier (PRESSO) were applied. A sensitivity analysis was conducted using the leave-one-out method. Results: The  In humans, up to 100 trillion bacterial cells from approximately 500 distinct species are present, and the gastrointestinal (GI) tract is the main habitat for >70% of those microbes . ImportaEvidence of a link between the gut microbiota and renal function is rare; however, very recently, the role of the gut microbiota in renal physiology and pathophysiology has been investigated ,12,13. TEntrobacteraceae and Pseudomonadaceae genera of the phyla Proteobacteria, Bacteroidaceae and Clostridiaceae and a decrease in species of Lactobacillaceae, Prevotellaceae and Bifidobacteriaceae [Lactobacillus and Bifidobacterium in patients with CKD may lead to a decrease in uric acid, blood urea nitrogen (BUN) [It has been reported that gut dysbiosis in CKD includes an increase in the species of eriaceae . In end-eriaceae . Moreoveen (BUN) ,27, seruen (BUN)  and estien (BUN) .Desulfovibrionaceae, Bacteroidaceae, Alcaligenaceae, Pseudomonadaceae, and Pasteurellaceae (producing urease) were much lower in ESKD patients than in normal individuals, while Bacteroidaceae relative abundance was higher in these patients [Bacteroides was positively correlated with the level of creatinine and dominant in ESKD patients [Faecalibacterium was positively associated with eGFR and dominant in controls [Bifidobacterium was negatively associated with BUN and creatinine [Anaerostipes spp., as well as an increased relative abundance of Akkermansia spp. and Lactobacillus spp. [While epidemiological studies are limited because they are not able to rule out the chance of residual bias, confounding factors and reverse causation, Mendelian Randomization (MR) analysis is able to compensate for these limitations, when applied to both phenotypes and genotypes . A recenpatients . Reportspatients , whereascontrols . Moreoveeatinine . These leatinine . Furtherlus spp. .Taking the abovementioned factors into account, we conducted an MR analysis. This method is able to circumvent the limitations of observational studies  by usingA two-sample MR study design was used. Summary statistics were obtained from the largest GWAS on microbiota and outcomes of interest. We applied methods to estimate the unbiased effect of eight microbiome genera on the risk of CKD and renal function parameters (eGFR) separately in non-DM and DM participants.Acidaminococcus [Anaerostipes [Bacteroides [Bifidobacterium [Desulfovibrio [Faecalibacterium [Lactobacillus [Megamonas [From all GWAS, published on different microbiota genera, we used data for genera which had ≥2 single-nucleotide polymorphisms (SNPs): nococcus , Anaerosrostipes ,34, Bactacterium ,34, Desufovibrio , Faecaliacterium ,34, Lactbacillus , Megamonegamonas . More inegamonas ,33,34.2. DM was defined as fasting glucose ≥126 mg/dl, with pharmacologic treatment for DM or self-reported treatment [R2 > 0.99) with the original SNP, was used as proxy when the original SNP was not available for outcomes.Genetic associations with renal function were obtained from the largest available extensively genotyped study based on a meta-analysis  . eGFR wareatment . In all reatment . For genreatment . Logistireatment . An SNP,p-value of the intercept term from MR-Egger [We combined the effect of 3 instruments using the inverse-variance weighted (IVW) method as implemented in the TwoSampleMR package running under R. We assessed heterogeneity using the Q value for the IVW method. To address the potential effect of pleiotropic variants on the final effect estimate, we conducted sensitivity analysis including weighted median (WM) and MR-Egger. Sensitivity analysis was conducted using the leave-one-out method. WM estimate, as the weighted median of the SNP-specific estimates, provides correct estimates as long as SNPs accounting for ≥50% of the weight are valid instruments. WM MR allows some variants to be invalid instruments provided at least half are valid instruments. It uses IVW and bootstrapping to estimate confidence intervals (CIs) . MR-EggeMR-Egger . Causal MR-Egger . AnalysiMR-Egger .Further, to assess heterogeneity between individual genetic variant estimates, we used the Q’ heterogeneity statistic  and the This investigation used published or publicly available summary data with no identification of the participants. No original data were collected for this manuscript. Ethical approval for each of the studies included in the investigation can be found in the original publications (including informed consent from each subject). The study conforms to the ethical guidelines of the 1975 Declaration of Helsinki.Anaerostipes was associated with a higher eGFR in the overall population ; they had also no significant effect on the risk of CKD .We also evaluated the impact of p > 0.322 for all comparisons). The analysis of MR-PRESSO did not highlight any outlier for all estimates. The pleiotropy test (for those genera with more than two SNPs), with very negligible intercept and insignificant p-value, also indicated no chance of pleiotropy for all our estimations . The results of MR-RAPS were identical with the IVW estimates, which again indicated no chance of pleiotropy. The leave-one-out method demonstrated that the links were not driven by a single SNP.The results of heterogeneity indicated no chance of heterogeneity . Both serum creatinine and BUN increased around 2.5-fold in the NPX group compared with the Sham group [Anearostipes spp. [Anaerostipes showed negative effects on kidneys parameters (BUN and creatinine) [Anearostipes [Anearostipes led to increased eGFR and improvement in renal function.butyrate . Increashe blood . Bacterihe blood . Bacterohe blood ,59. Butyhe blood . Bacterihe blood . It has he blood . There ivailable . C57BI/6am group . The divpes spp. . NPX micatinine) . Furtherrostipes . The resDesulfovibrio genus includes sulfate-metabolizing bacteria that reduce the sulfites and sulfates obtained from the diet and the sulfated mucopolysaccharides found in mucin, leading to the generation of hydrogen sulfide, a cytotoxic compound [Desulfovibrio spp. produced urease, whose relative abundance was much lower in ESKD patients than in the control group [Desulfovibrio spp. clearly flourish in an inflammatory environment [The compound . A recenol group . It was ironment , and harironment ,64,65.Desulfovibrio spp. on kidney function. The main theory centres on the fact that SCFAs such as butyrate are metabolized by Desulfovibrio. Then, Desulfovibrio uses sulfites and sulfates as terminal metabolic electron acceptors, leading to the generation of hydrogen sulfide, which has a wide range of cytotoxic effects. Among these, hydrogen sulfide is a potent inhibitor of the oxidation of SCFAs in cells, thereby completing a vicious circle of mutually exclusive metabolic interactions between host and this bacterium [Several mechanisms have been proposed for the detrimental effect of acterium .Desulfovibrio and Anearostipes with eGFR and CKD. It should be noted that different mechanisms and factors in diabetic and non-diabetic individuals may explain why we did not find any association in the diabetic subgroup [With regard to diabetic populations, we did not find an association of subgroup . Oxidatisubgroup . CKD as subgroup . It coulOur analysis has strengths and limitations. MR is a powerful tool for the detection of causal impacts, which is an advantage with respect to observational studies. Moreover, we benefited from the largest GWAS on exposure and outcomes. However, the MR analysis has limited statistical power. Therefore, a lack of findings in our analysis might be due to small causal effects that were not detectable in our study. Our study used genetic instruments and estimation of genetic associations using GWAS that were mainly based on a sample of European ancestry. It is known that the microbiota genera vary by ethnicity and other factors. Therefore, our results cannot be extrapolated to other ethnicities. We could not test for subgroup analysis  and diet–microbiome interactions, but causal effects should be generally consistent. Second, the 16S ribosomal RNA gene sequencing used by most microbiota GWAS usually only permits resolution at the genus level rather than at a more specific level, so we cannot rule out the possibility that some specific species or strains are associated with renal function. Moreover, for future studies, it is important to consider the ratio between two taxa or dysbiosis of the gut microbiota. Additionally, the gut microbiota may also be influenced by other factors, such as age and season. However, the gut microbiota is thought to be stable especially after early childhood, and the main force determining its composition consists of long-term dietary habits. As such, our findings may be more relevant to the effects of the gut microbiota in adolescence or adulthood. We would like to call for more GWAS with greater sample size to include other bacterial genera . Our study is also limited by the current understanding of the gut microbiota. A hypothesis-driven study testing epidemiologically established associations would have been preferable. However, this was precluded by the lack of knowledge as to the function of each microbiome constituent and the lack of large epidemiological studies. However, for both genera which had an effect on renal function, there were experimental studies enabling us to discuss possible mechanisms.Desulfovibrio and Anearostipes spp. on renal function markers. Further studies using multiple robust instruments are needed to confirm these results.In conclusion, our study highlighted the association of


Downstream remodeling of the spiral arteries (SpA) decreases utero-placental resistance drastically, allowing sustained and increased blood flow to the placenta under all circumstances. We systematically evaluated available reports to visualize adaptation of spiral arteries throughout pregnancy by ultra-sonographic measurements and evaluated when this process is completed.A systematic review and meta-analysis of spiral artery flow (pulsatility index (PI), resistance index (RI) and peak systolic velocity (PSV)) was performed. English written articles were obtained from Pubmed, EMBASE and Cochrane Library and included articles were assessed on quality and risk of bias. Weighted means of Doppler indices were calculated using a random-effects model.p < 0.001) and 0.39  in the second trimester and to 0.49  and 0.36  in the third trimester, respectively. In parallel, PSV altered from 0.22 m/s (95% CI: 0.13–0.30 m/s) to 0.28 m/s  and to 0.25 m/s  in the three trimesters. In absence of second and third trimester Doppler data in complicated gestation, only a difference in PI was observed between complicated and healthy pregnancies during the first trimester . Although individual studies have identified differences in PI between SpA located in the central part of the placental bed versus those located at its periphery, this meta-analysis could not confirm this (p = 0.349).In healthy pregnancies, PI and RI decreased from 0.80 (95% CI: 0.70–0.89) and 0.50 (95% CI: 0.47–0.54) in the first trimester to 0.50 (95% CI: 0.45–0.55, This review and meta-analysis concludes that an observed decrease of SpA PI and RI from the first towards the second trimester parallels the physiological trophoblast invasion converting SpA during early gestation, a process completed in the midst of the second trimester. Higher PI was found in SpA of complicated pregnancies compared to healthy pregnancies, possibly reflecting suboptimal utero-placental circulation. Longitudinal studies examining comprehensively the predictive value of spiral artery Doppler for complicated pregnancies are yet to be carried out. The most important functions of the placenta are adequate exchange of oxygen, nutrients and waste products between fetal and maternal circulation . At 12–1The etiology of placenta syndrome (PS), a collective term for placenta-insufficiency related disorders, is poorly understood. Inadequate remodeling of SpA in PS-pregnancies potentially affects the development of the placenta as high velocity blood flow could cause mechanical stress on the tissue, or could influence development due to perfusion-reperfusion injury , 7, 10.The golden standard for the detection of absent or aberrant SpA remodeling is postpartum histopathological examination of the placenta, however this is behind time to affect clinical decision making in endangered gestation . PreviouThis systematic review and meta-analysis on the remodeling of SpA flow velocimetry throughout human pregnancy is part of a large series of meta-analysis on physiological and pathophysiological adaptation of relevant indices in pregnancy –18. The The electronic databases Cochrane library 1997–2019), Pubmed (1946–2019) and EMBASE (1974–2019), were searched for relevant articles evaluating SpA Doppler measurements. The keywords used in the literature search were “spiral artery” or “spiral remodeling” in combination with “Doppler” or “color Doppler”, or “Doppler velocimetry”, or “Doppler sonography” or “spiral artery Doppler”, see Table 997–2019,Additionally, the reference list of all selected primary articles were examined for potential citations not captured by the initial search. The search was limited to papers published in English until April 2019.The first selection, based on title and abstract, was performed independently by two investigators (VS and LE). In case of discrepancy, agreement was reached by consensus. The second selection was also performed by two investigators (VS and LE) independently, based on the full text. After checking the manuscripts and crosschecking their reference lists, the final selection of studies was made. Inclusion criteria were studies on singleton pregnancies in which absolute values for SpA Doppler measurements were documented at any gestational age during pregnancy. Studies with both nulliparous and multiparous women were included; no limitations were set on maternal characteristics. Moreover, both transvaginal and transabdominal measurements were included as well as measurements of central localized and peripheral localized SpA. We initially wanted to include only articles describing SpA measurements in healthy and hypertensive pregnancies. Given the limited amount of articles describing SpA measurements in hypertensive pregnancies, we decided to include all articles describing SpA measurements in complicated pregnancies. Therefore, complicated pregnancies in the included articles were represented by PE, preterm labor, FGR, missed abortion, miscarriage and placental abruption.In some studies, absolute values of Doppler indices could not be extracted as a result of data being reported in graphs or figures. We considered the use of, for example, a plot digitizer to extract the data out of graphs or plots. The accuracy of the assessment was in some cases questionable, and in other cases, not all graphs and plots showed 95% confidence interval (CI) lines, making it impossible to calculate a reliable standard deviation (SD). Therefore, adding a plot digitized estimated SpA value would have a great impact on the trustworthiness of the pooled mean. Therefore, we refrained from generating extra data using the plot digitizer as it may be less accurate compared to the other included articles.Two reviewers (VS and LE) both extracted the data and assessed the risk of bias. In case of discrepancy in assessment of the risk of bias, agreement was reached by consensus. The following data was extracted from each included study: 1) Study design; 2) Sample size; 3) Probe; 4) Doppler method 5) MHz of the probe; 6) Gestational age during measurement; and 7) Doppler velocimetry outcomes  tool  was modiScoring of each criteria occurred as insufficient [−] or sufficient [+]. In case an item was not applicable for the study, a question mark [?] was used. Based on the number of [+], the total score as a percentage was calculated. Articles scoring < 30% were defined as low quality, between 30 and 60% as moderate quality, and > 60% as high quality.2 statistic. I2 can differentiate between true heterogeneity and sampling variance [p < 0.05.The obtained SpA measurements were categorized into three pregnancy trimesters and divided by healthy versus complicated pregnancies and central versus peripheral SpA. Some articles performed two or more SpA measurements in one pregnancy trimester in the same population, explaining the repeated presence of one article in the forest plots. The analysis was conducted by using the R Project for Statistical, R version 3.4.0 with ‘meta’package version 2.0–0 . In few variance . Differen = 409), Embase (n = 143) and Cochrane (n = 20). No restrictions were set on publication status. We were unable to add any extra articles that were not identified by the literature search after examination of the reference lists. After removal of duplicates (n = 92), 467 articles were screened for eligibility based on title and abstract. Of these, 442 records were discarded based on title and abstract because these papers investigated other study outcomes, included animal studies, were reviews or were written in a non-English language that the investigators did not understand. Of the remaining 25 records, full-text articles were retrieved and examined in detail for eligibility. Thirteen additional studies were excluded, due to only estimates in figures were presented of spiral artery Doppler indices and no absolute values. We contacted the authors of whom their correct contact information was reported in the articles, to see whether they were able to provide additional data. Unfortunately, only two authors replied that they could not help us, since the research was performed before digitalization and it was not possible to acquire the absolute data of the dated investigations.Figure n = 7) and cross-sectional studies (n = 5). SpA measurements were obtained using transvaginal (n = 6), transabdominal ultrasound (n = 5) or both (n = 1) with probes between 3.5 to 12 MHz. All articles evaluated SpA Doppler during healthy pregnancy. Besides, five articles investigated the differences in SpA Doppler measurements between complicated and healthy pregnancies. There were no articles found that examined SpA Doppler measurements in complicated pregnancies only. Detailed information on the characteristics of included studies is presented in Table A total of 12 studies published in English met the eligibility criteria and were included in this systematic review and meta-analysis. The selected articles were longitudinal prospective (n = 5) [n = 4).The quality assessment according the modified QUIPS tool per domain is summarized in Table n = 5) , 27, 28.n = 5) , Kurjak n = 5)  and Matin = 5)  were scon = 5) , Hsieh en = 5)  and the n = 5)  scored hStudies reported Doppler indices at different gestational ages. If one mean SpA Doppler parameter was reported with an interval of gestational weeks as time of measurement, we included the mean of the interval as time of measurement in the meta-analysis . Doppler indices were combined based on measurements during first trimester (0–14 weeks), second trimester (15–27 weeks) or third trimester (28–40 weeks).n = 9) in the first trimester to 0.50  (n = 5) in the second trimester and to 0.49  (n = 2) in the third trimester (Table Ten studies explored the PI at certain (between 5 and 39.5 weeks of gestation) time-points during healthy gestation. The weighted mean of PI decreased from 0.80 (95% CI: 0.70–0.89) (n = 7) in the first trimester to 0.39  (n = 4) in the second trimester and 0.36  (n = 2) in the third trimester . The weighted mean of RI showed a decrease from 0.50 (95% CI: 0.47–0.54) (n = 3) in the first trimester to 0.28 m/s  (n = 2) in the second trimester and remained 0.25 m/s  (n = 2) in the third trimester . The weighted mean of PSV increased from 0.22 m/s (95% CI: 0.13–0.30) (p = 0.320). Results of Ozkaya et al. showed mean SpA PI in 16 pregnancies with adverse outcome  was 0.97 ± 0.51 (mean ± SD) vs 0.82 ± 0.39 in 84 women with normal pregnancy outcome. The difference in PI between both groups was non-significant. When pooling these data, a statistically significant difference between complicated and healthy pregnancies in the first trimester was found . Likewise, Ozkaya et al., found a non-significant difference in SpA RI between women with adverse pregnancy outcome compared to women with normal outcome (0.60 ± 0.37 vs 0.54 ± 0.10). Makikallio et al. described SpA RI during 6, 8, 9 and 11 weeks of gestation in 10 women having PE or preterm labor and compared them with 31 control pregnancies . During the whole study period, no significant differences between both groups were observed. Equally, our pooled meta-analysis data showed no significant differences in RI between complicated and healthy gestation throughout the first trimester . No data was available on RI measurements in second or third trimester pregnancies.Both articles, along with a third article , investip < 0.001; peripheral SpA RI: 0.38 ± 0.21 vs central SpA 0.32 ± 0.18, p < 0.001). No significant differences were found in the PSV between the central and peripheral spiral arteries (p > 0.05). Studies of Hsieh et al. concluded that PI and RI values of central placenta bed SpA seemed to be lower, although this finding was statistically non-significant due to the small number of subjects in their sample.Differences between central and peripheral SpA Doppler indices during the second trimester were investigated in two studies , 23. Sigp = 0.349)  , compared to 0.53 (95% CI: 0.53–0.62) in peripheral SpA (84) Fig. .Fig. 9FIn healthy gestation, we observed a consistent decrease of PI and RI in SpA from the first to the second trimester, after which no relevant changes towards the third trimester were seen. Transformation of adrenergic-sensitive high-resistance into adrenergic-insensitive low-resistance SpA ensures unrestricted blood flow into the placenta , 35. DurThe exact time at which high oxidative maternal blood starts to flow into the intervillous space ranges from 5 to 12 weeks of gestation and is prevented earlier by the formation of endovascular trophoblast ‘plugs’ in the distal segment of the SpA , 21, 24.Defective remodeling of SpA can lead to a premature onset of maternal, high-flow and -oxidative blood into the placenta, generating mechanical and biochemical trophoblastic damage and increased apoptosis . In PS-pThree of the included studies reported PI/ RI in healthy and complicated pregnancies , 27, 28.Pijnenborg et al.  found thThe use of Doppler ultrasonography offers a non-invasive technique for the measurement of SpA flow . NonetheLost to follow up was not documented in the longitudinal designs, possibly influencing the constructed meta-regression curves. Most studies only performed measurements during first trimester, and those studies used transvaginal probes, whereas transabdominal probes were used in studies measuring during the second and third trimester. However, Marchi et al.  showed tDespite the fact that upstream uterine measurements are closely related to downstream SpA measurements , it is a


We performed a comprehensive meta-analysis to precisely estimate of the impact of the hOGG1 Ser326Cys polymorphism on gynecologic cancer susceptibility. Electronic databases including PubMed, Embase, WanFang, and the China National Knowledge Infrastructure were searched for relevant studies. Odds ratios (ORs) with corresponding 95% confidence intervals (CIs) were determined to assess the strength of the association. Fourteen studies with 2712 cases and 3638 controls were included in the final meta-analysis. The pooled analysis yielded a significant association between the hOGG1 Ser326Cys polymorphism and overall gynecologic cancer susceptibility . A significantly higher gynecologic cancer risk was found for the European population , but not in the Asian population. The stratified analysis by cancer type revealed endometrial cancer was significantly associated with the hOGG1 Ser326Cys polymorphism . In conclusion, the hOGG1 Ser326Cys polymorphism was associated with higher overall gynecologic cancer susceptibility, especially for endometrial cancer in the European population.The association between the  Gynecologic cancer, including cervical, ovarian, and endometrial cancer, is the major cause of cancer-related deaths in women worldwide. Among them, cervical cancer is the fourth leading cause of deaths with an estimated 311,000 deaths in 2018 worldwide, followed by 185,000 deaths from ovarian cancer, and over 89,000 deaths from endometrial cancer [DNA damage can initiate the genetic instability that drives cancer development ,4. DNA dhOGG1 gene is located on chromosome 3p26.2 and encodes a glycosylase that catalyzes the excision of 8-oxoguanine adducts from damaged DNA [hOGG1 knockout mice, 8-oxoguanine was found to accumulate in the genomic DNA and carcinoma developed spontaneously [hOGG1 gene have been confirmed to exert different activities in the repair of 8-oxoguanine in the complementation assay of defective Escherichia coli variant, carrying, for example, the hOGG1 Ser326Cys (rs1052133C>G) polymorphism [hOGG1 Ser326Cys polymorphism and gynecologic cancer risk. However, there are still many inconsistencies. An increased risk was found by some studies [hOGG1 Ser326Cys polymorphism and the susceptibility to gynecologic cancer. To date, only one meta-analysis evaluated the association with overall gynecologic cancer using subgroup analysis and included only five studies published before September 2014 [hOGG1 Ser326Cys polymorphism and susceptibility to gynecologic cancer.The aged DNA . In hOGGaneously . Furthermorphism . To date studies , but was studies . Thus, aber 2014 . Nonetheber 2014 ,18. ThusRelevant studies were retrieved from electronic databases including PubMed, Embase, WanFang, and the China National Knowledge Infrastructure using the following search terms: ‘hOGG1 or OGG1’, ‘polymorphism or variant or variation’, and ‘cancer or tumor or carcinoma’. We also evaluated previously published meta-analyses and review articles for other relevant studies. The language was limited to English and Chinese. All the relevant articles were searched before August 1, 2020.hOGG1 Ser326Cys polymorphism and susceptibility to gynecologic cancer and (2) available information about genotype and allele frequency for genetic model analysis. The exclusion criteria were as follows: (1) no case–control studies; (2) reviews, case reports, meta-analyses, and letters and (3) studies with insufficient data. In addition, studies with genotype frequencies in the control groups deviating from Hardy–Weinberg equilibrium (HWE) were also included in our meta-analysis if further evidence showed that other polymorphisms did not deviate from HWE.The inclusion criteria were set as follows: (1) case–control study design evaluating the association between the According to the inclusion and exclusion criteria listed above, two authors independently extracted information from all eligible publications and any disagreements were resolved through discussions with other authors. The following information was collected from each included study: first author’s surname, year of publication, country, ethnicity, cancer type, source of control, genotyping method, and genotype frequency. Different ethnicities were categorized as either European or Asian population and different sources of control were defined as hospital-based (HB) and population-based (PB). The quality of each study was evaluated using quality assessment criteria as outlined in previous studies . The quaZ-curve crossed the TSA monitoring boundary or exceeded the required information size, firm evidence had been reached. Otherwise, more studies were needed [Trial Sequential Analysis Viewer (TSA)  was used to perform the analysis using the data retrieved from the studies. The parameters were set as followed: 5% type I error, 20% relative risk reduction, and 20% type II error . If the cumulative e needed .hOGG1 Ser326Cys polymorphism and the susceptibility to gynecologic cancer was evaluated by calculating the odds ratios (ORs) with their corresponding 95% confidence intervals (CIs). The pooled ORs were obtained for the homozygote model (Cys/Cys vs. Ser/Ser), heterozygote model (Ser/Cys vs. Ser/Ser), dominant model (Ser/Cys + Cys/Cys vs. Ser/Ser), recessive model (Cys/Cys vs. Ser/Cys + Ser/Ser), and the allele model (Cys vs. Ser). Stratified analyses were also carried out for ethnicity, cancer type, source of control, and genotyping method. The Chi-square-based Q test was used to assess the degree of heterogeneity among studies, and a P-value < 0.10 was regarded as significant. If no heterogeneity existed with P > 0.10, the fixed-effect model was used [https://www.gtexportal.org/home/) was used to evaluate the correlation between the hOGG1 Ser326Cys polymorphism and levels of mRNA expression of hOGG1. All statistical tests were performed using STATA version 11.0 .The Chi-square test was applied to assess the HWE in the control groups. The strength of the association between the was used . Otherwiwas used . Sensitiwas used . Only thhOGG1 Ser326Cys polymorphism [A total of 498 relevant articles were retrieved from the electronic database search. Overall, 480 publications were excluded after title and abstract screening. Of the remaining 18 publications, one study was excluded because of insufficient genotype information for data analysis ; one stumorphism , and twomorphism ,36. Overmorphism ,37. The morphism ,17,19,25hOGG1 Ser326Cys polymorphism and increased susceptibility to gynecologic cancer . In the stratified analysis by ethnicity, a significantly increased gynecologic cancer risk was found in the European population . Thus, there was no significant change in the ORs and 95% CIs after recalculations by sequentially removing one single study, suggesting the stability of our results.Substantial heterogeneities were observed among all studies for the association between the P=0.182, heterozygous model: P=0.228, recessive model: P=0.149, and dominant model: P=0.252), except for the allele model (P=0.035).The funnel plot and Egger’s liner regression test were used to detect a potential publication bias. As shown in Z-curve for the hOGG1 Ser326Cys polymorphism was evaluated using TSA analysis, and showed the pooled studies failed to reach the required information size, which suggested that the cumulative boundary was not achieved and further studies with larger sample sizes would be required to verify these associations.As shown in We selected 0.2 as the FPRP threshold. As listed in hOGG1 Ser326Cys polymorphism on the mRNA expression of hOGG1 using the GTEx web tool. The 326Cys allele was significantly associated with higher expression of hOGG1 in the cultured fibroblasts cells (We further assessed the impact of the ts cells .hOGG1 Ser326Cys polymorphism, the most studied hOGG1 polymorphism, was found to influence protein activity, and thus, contribute to carcinogenesis [hOGG1 Ser326Cys polymorphism and gynecologic cancer susceptibility. However, the results have been conflicting and inconclusive.Gynecological carcinogenesis is still poorly understood. It is a complex event influenced by environmental and genetic factors, as well as by gene–environment interactions. In recent years, genetic factors are increasingly believed to be important contributors to carcinogenesis, and, the imbalance between DNA damage and repair ,39. Overogenesis . CurrenthOGG1 Ser326Cys polymorphism and gynecologic cancer susceptibility. Overall, the current meta-analysis included 14 studies with 2712 cases of gynecologic cancer and 3638 controls and observed a significant association between the hOGG1 Ser326Cys polymorphism and overall gynecologic cancer susceptibility. Furthermore, stratified analysis by ethnicity indicated that hOGG1 Ser326Cys polymorphism was associated with increased gynecologic cancer risk in the European population, but not in the Asian population. Moreover, in the stratified analysis by cancer type, a statistically significant association was identified among ovarian cancer and endometrial cancer, but not with cervical cancer.To the best of our knowledge, this is the first meta-analysis focusing on the association between the hOGG1 Ser326Cys polymorphism and gynecologic cancer susceptibility [hOGG1 Ser326Cys polymorphism was associated with overall gynecologic cancer susceptibility in a homozygous model [hOGG1 Ser326Cys polymorphism and endometrial cancer [hOGG1 Ser326Cys polymorphism was associated with an increased risk of endometrial cancer. As expected, there have been a relatively small number of studies that have focused on this specific association. However, we collected all the available studies and provided a comprehensive evaluation of their relationship. Further, we applied TSA to verify the above results, which suggested that additional studies with larger sample sizes would be needed to confirm these associations.Two previous meta-analyses investigated the association between the tibility ,43. Bothus model . As for hOGG1 Ser326Cys polymorphism was statistically associated with the susceptibility of ovarian cancer and endometrial cancer, but not of cervical cancer. Furthermore, we conducted a statistical analysis stratifying these three types of gynecologic cancer, as shown in The stratified analysis by cancer type indicated that the Several limitations of this meta-analysis should be addressed. First, lack of original data limited any additional estimation of gene–gene and gene–environment interactions. Second, the number of cases and controls enrolled in the current meta-analysis was relatively small. Third, notable heterogeneities were observed for several models, thus a random-effect model was applied.hOGG1 Ser326Cys polymorphism was associated with an increased risk of overall gynecologic cancer susceptibility, especially for endometrial cancer in the European population. More eligible case–control studies are necessary to further confirm our findings.In conclusion, our study revealed that the


Background and objectives: The prognostic role of a disintegrin and metalloproteinase (ADAM) 17 has been widely assessed in gastric cancer. However, the results are inconsistent. We performed a meta-analysis to evaluate the prognostic significance of ADAM17 and its association with clinicopathological parameters. Methods: The databases of PubMed, Web of Science, and Embase were searched for relevant articles published up to April 2020. The reported hazard ratios (HRs) and odds ratios (ORs) and their corresponding 95% confidence intervals (CIs) were pooled to evaluate the strength of the association. Stata 12.1 was used to perform statistical analyses. Results: Seven studies, including 1757 patients, were screened for the meta-analysis. Compared with the high ADAM17 expression group, the pooled HR was higher in the low ADAM17 expression group . High ADAM17 expression was also related to the tumor node metastasis (TNM) stages , lymph node metastasis , and ages  of the gastric patients. Conclusions: This meta-analysis revealed that ADAM17 is a significant biomarker for poor prognosis in gastric cancer. Gastric cancer (GC) is one of the most common malignancies and the second leading cause of cancer-related death . There wThe ADAMs  are a multifunctional family of proteins, many of which are associated with the formation and progression of cancer ,7. SeverThe association between ADAM17 and gastric cancer has been described in several studies, but the significance of ADAM17 for the prognosis of the cases is inconsistent among reports ,25,27,28Literature was searched in PubMed, Embase, and Web of Science until April 2020. The main search keywords are as follows: “stomach tumor” or “stomach cancer” or “stomach carcinoma” or “gastric tumor” or “gastric cancer” or “gastric carcinoma” and “ADAM17 protein” or “TNF-alpha converting enzyme” or “TNF alpha converting enzyme” or “ADAM-17” or “Tumor necrosis factor-alpha converting enzyme” or “Tumor necrosis factor alpha converting enzyme” or “TNF-alpha convertase” or “ADAM-17 protein” or “ADAM17 protein”. In addition, references in original studies and reviews were manually searched.Studies were included if they met the following criteria: (1) the participants were patients with GC; (2) they reported ADAM17 level and hazard ratios (HRs) with 95% confidence intervals (CIs) of patients or provided the ADAM17 levels and clinical information; (3) ADAM17 had an explicit inspection method . Studies would be excluded if (1) the studies involved the same population that the studies have incorporated; (2) the articles were case reports, letters, comments, reviews, or conference summaries.The two authors (P.N. and M.Y.) independently extracted data. It included age, gender, the first author of the study, follow-up time, sample size, publication year, HRs, 95% CIs for overall survival (OS). If there was the same cohort of patients in several studies, the article with sufficient information, long follow-up time, and large sample size was selected. The two investigators (P.N. and M.Y.) independently assessed the quality of the selected literature using the Newcastle-Ottawa-Scale (NOS) . The scoThe HRs and corresponding 95% CIs were used to estimate the effect of ADAM17 expression on survival rates. The HRs and 95% CIs would be directly extracted, if they were reported in the original studies. Otherwise, the HRs were estimated using the methods described by Parmaretal et al. . The oddI2 statistics and Cochran Q were used to test the heterogeneity [I2 statistics was 25%, 50%, 75%, considered as low, moderate, and high heterogeneity, respectively. In addition, p < 0.10 was considered statistically significant for the Q statistic [I2 statistics was more than 50% or the p-value was less than 0.1 for the Q statistic; otherwise, a fixed effects model was utilized. We conducted a sensitivity analysis to evaluate the stability of the results by excluding a single study at one time. The Begg’s test and Egger’s test were used to assess publication bias [The ogeneity . The valtatistic . The ranion bias . We usedWe identified 267 potentially relevant articles from PubMed, Web of Science, and Embase databases for the meta-analysis, and 51 of them were excluded because of duplicates. After screening the titles and abstracts, 203 articles were removed, and 13 articles were retrieved for full-text assessment. Finally, seven with a total of 1757 participants were included in this meta-analysis ,27,28,33The main characteristics of the eligible studies are in I2 = 18.1%; p = 0.299). Our study indicated that the poor prognostic effect of ADAM17 in GC was significant. Compared with the high ADAM17 expression group, the pooled HR was higher than in the low ADAM17 expression group  (A fixed-effect model was adopted in the study considering the low level of heterogeneity (= 0.299) . I2 = 84.1%; p = 0.000), lymph node metastasis , and age   and sex .The elevated expression of ADAM17 was significantly associated with such characteristics as higher TNM stage  . There ap = 1.000) and the Egger’s test (p = 0.493). Sensitivity analysis showed that all of the estimated pooled HRs corresponding to the omission of each study was inside the 95% CI of the HR estimated from all studies in the overall, indicating that the results of the present study were stable. No significant publication bias was found by the Begg’s test  status and was higher in GC patients with intact cagPAI strains [Some studies showed that biological and environmental factors, pression ,41,42,43al cells . In addiA previous meta-analysis showed ADAM17 was correlated with TNM stages and distant metastasis, but not correlated with cancer grade of stomach cancer . In our The advantages of this meta-analysis are as follows: Firstly, this study performed more comprehensive analysis of the data, including prognostic value and clinicopathological parameters. Secondly, all the data was extracted directly from the articles. However, among the included studies, there were some limitations. Firstly, the definition of level of ADAM17 expression is not consistent in the included studies. Secondly, most of the articles were from China and the results were applicable to Asian populations. Thirdly, our study included few articles, and the results of our study need to be validated by large prospective cohorts.Our study shows that lower ADAM17 levels are correlated to the longer OS time in GC and ADAM17 was associated with TNM stage, lymph node metastasis, and age. ADAM17 is a significant biomarker for poor prognosis in gastric cancer.


Peripherally inserted central catheter (PICC) is closely related to venous thromboembolism (VTE). It is a common complication of hospitalized patients, and its etiology is complex. How to prevent the occurrence of VTE is the focus of clinical work. In order to provide basis for individual prevention and accurate treatment of VTE, the purpose of this study was to explore the relationship between ABO blood group and the risk of VTE in cancer patients associated with PICC through meta-analysis.Electronic databases, including Embase, Cochrane Library, Pubmed, Chinese databases SinoMed, Chinese National Knowledge Infrastructure (CNKI), Chinese Scientific Journals Database (VIP), and Wanfang Data, were searched for case-control studies of ABO blood group and PICC-related VTE in cancer patients. The languages were limited to Chinese and English. Two reviewers were responsible for the selection of the study, the extraction of data and the evaluation of the quality of the research. All statistical analyses were carried out with Review Manager 5.3 and Stata 14.0.The results of this meta-analysis would be published in peer-reviewed journals.This study provided evidence to support the relationship between ABO blood group and the risk of PICC-related VTE in cancer patients.DOI 10.17605/OSF.IO/6DPFG The incidence of PICC-related venous thromboembolism (VTE) significantly increased, and became one of the most common factors of PICC catheter wall plug.–10Peripherally inserted central venous catheter (PICC) is widely used in chemotherapy and nutritional support for tumor patients.,12 A number of studies proved that ABO blood group has an obvious impact on coagulation function, because ABO blood group is the main determinant of plasma von Willebrand factor (vWF) level.–15Many studies revealed that patients with non-O blood group have an increased risk of VTE.Overall, about 70% of the changes in vWF/F VIII plasma levels is determined by heredity, of which about 30% is related to individual ABO blood type. It is worth noting that about 25% of individuals with non-O blood groups has high levels of vWF. Many scholars further explore the possible clinical significance of this biological relationship. For example, whether ABO blood type may be a factor that affects the risk of bleeding or arterial / VTE thrombosis, etc. Gándara et al observed that the incidence of VTE is high in people with non-O blood group, and considered that non-O blood is a risk factor for VTE. Isabel et al discovered that high VII level and non-O blood group are independent risk factors for VTE, and should be considered in the evaluation of thrombotic hemophilia.Previous studies displayed that people with nontype 0 blood have a remarkable higher risk of VTE and coronary heart disease than those with type O blood.,20 In this paper, in order to better clarify the possible relationship between ABO blood group and PICC-related VTE in cancer patients, and to provide a basis for individualized prevention and accurate treatment of ABO blood group and PICC-related VTE in cancer patients, we collected a case-control study on the relationship between ABO blood group and the risk of PICC-related VTE in cancer patients for meta-analysis.The common risk factors of PICC-VTE are well known, but there are few studies on the relationship between ABO blood group and them. At present, the relationship between ABO blood group and PICC-related VTE in cancer patients is still controversial.22.1The protocol was registered in Open Science Framework (registration number: DOI 10.17605/OSF.IO/6DPFG). This systematic review and meta-analysis will be reported inconformity with the preferred reporting items for systematic reviews and meta-analysis protocols (PRISMA-P) 2015.2.2The review does not involve the assessment of patients’ individual information or rights, so there is no need to obtain approval from an ethical institution.2.3Studies would be included in this meta-analysis based on following criteria:1.Study types: All case-control studies related to ABO blood type and PICC-related VTE in cancer patients susceptibility should be included. The publication of literatures is unrestricted.2.Participant type: Tumor patients implanted with PICC should be included in the meta-analysis. Not subject to age, sex, or national restrictions.3.Outcome: PICC-related VTE risk comparisons.2.4According to the following criteria, studies should be excluded from the meta-analysis: Conference summaries, incomplete data studies, repeated published studies, and case series.2.5Embase, Cochrane Library, Pubmed, Chinese databases SinoMed, Chinese National Knowledge Infrastructure (CNKI), Chinese Scientific Journals Database (VIP), and Wanfang Data were searched. The details of PubMed's search strategy are shown in Table 2.62.6.1Two researchers read the questions and abstracts alone to screen relevant literatures. By reading the full text, the research design, inclusion and exclusion criteria, comparison and judgment of the research results of the relevant literature were evaluated. According to the inclusion and exclusion criteria, choosing whether to include the file in the meta-analysis. If there is a dispute, the third researcher would re-evaluate and decide whether to include the analysis. The screening flow chart of this study is displayed in Figure 2.6.2We extracted data from literatures that conform to the meta-analysis. The data includes the first author, year of publication, country, race of each study population, number of cases and controls, sex, average age, ABO blood type, and number of cases of thrombosis, etc.2.6.3 According to the evaluation content of Newcastle-Ottawa scale scale, they were divided into 8 items and 9 points, and each score was 1 point.Two researchers independently applied Newcastle-Ottawa scale to evaluate literatures included in the analysis, and adopted third-party opinions if there exist any differences.(1)The choice of the study population. Is the case selection appropriate? Is it representative? Is the choice of the control group the same population as the case group? How to determine the control group? One point per point, and a total of 4 points.(2)Comparability among groups. Does the study design and statistical analyses are taken into account the comparability between the case group and the control group, and whether the confounding factors are controlled? Age, race, region, gender, and sample structure were all regarded as important confounding factors, and those who controlled 1 item scored 1 point, with a total of 2 points.(3)The determination of exposure factors. How to determine the exposure factors? Is the same method used to determine the exposure factors in the case group and the control group? What is the nonresponse rate of the two groups? One point per point, and a total of 3 points.Literatures with a total score of more than 6 are considered to be high quality.2.6.4The research on the defects of the original data. We contacted the author by email and asked for the original data. If the original data is not available, then we would analyze the existing data.2.72 test (the test level was α = 0.1). Meanwhile, combining with I2 to quantitatively judge the size of heterogeneity. If there was no statistical heterogeneity among the results, the fixed effect model was adopted for meta-analysis. If there is statistical heterogeneity among the results of each study, the source of heterogeneity is further analyzed. After excluding the influence of obvious clinical heterogeneity, random effect model is adopted for meta-analysis. The level of meta-analysis was set as α = 0.05, and the obvious clinical heterogeneity was treated by subgroup analysis or sensitivity analysis, or only descriptive analysis.RevMan 5.3 (provided by Cochrane Collaboration) and Stata 14.0  were applied for statistical analysis. Odds ratio is the statistic of effect analysis, and each effect dose provides its 95% confidence interval. The heterogeneity among the included results was analyzed by χ2.8According to patient race, sample size, tumor type and so on, we carried out subgroup analysis.2.9Through the study of large weight of elimination effect, the sensitivity analysis was conducted to test the stability of the results of meta-analysis.2.10,25 In addition, publication bias was further quantitatively evaluated by Egger and Begg test.If more than 10 studies are included, a funnel chart would be applied to assess the report bias.2.11We utilized Grading of Recommendation Assessment, Development and Evaluation method to evaluate the evidence quality of the obtained results.3 However, many PICC complications can be observed in tumor patients, of which thrombus is an important consequence.,29 Tumor is a high risk factor for thrombosis, and the risk of thrombosis in tumor patients is 4 to 7.5 times higher than that in the general population.,10PICC provides an effective venous channel for patients undergoing chemotherapy and relieves the pain caused by repeated venipuncture.,31 that is a polymer with different molecular weight secreted by endothelial cells and megakaryocytes. As a protective carrier of coagulation factor VIII, it can mediate the adhesion between platelets and subendothelial collagen, and the aggregation between platelets and platelets, thus promoting thrombosis. VWF would be cleaved under the catalysis of metalloproteinase ADAMTS13 and eventually become a nonfunctional proteolytic fragment. The mechanism of ABO antigen system that affects thrombosis may be that its antigen exists on vWF through oligosaccharide chain. Different oligosaccharide chains of different blood groups affect the steric hindrance and charge effect of vWF structure, the effect of metalloproteinase ADAMTS13, and the rate of cleavage.,33 Furthermore, vWF initially existed in the form of a large complex, and when the endothelium was damaged, it folded tightly. The stacked vWF would become loose under the action of blood flow, exposing the acting groups and producing biological activity. The difference of oligosaccharide chain may affect the change rate of vWF structure.ABO blood group system is closely related to the level of vWFEarly prevention and intervention strategies are very important to reduce the incidence of PICC-related VTE for cancer patients. Determining the genetic composition of PICC-related VTE in cancer patients is a key area of VTE research. The identification of VET-related genes affects the understanding of its molecular and mechanism level, treatment and prevention. Therefore, in order to identify high-risk patients and design targeted treatment strategies to prevent serious complications in the future, we explored the ABO blood group. At present, although there are studies on the relationship between ABO blood group and PICC-related VTE in cancer patients risk. There is no systematic assessment of the cumulative evidence of this connection. We will conduct a systematic review and meta-analysis to clarify the relationship between ABO blood group and PICC-related VTE in the susceptibility of cancer patients.Data collection: Qiang ZhangFunding support: Qiang ZhangLiterature retrieval: Fangjing LiuSoftware operating: Fangjing LiuSupervision: Li Rao and Jifang SongWriting – original draft: Qiang Zhang, Fangjing Liu and Li RaoWriting – review & editing: Li Rao and Jifang Song


To date, the fundamental pathophysiology underlying the occurrence and progression of psoriasis are still unanswered questions. Genome-wide association surveys have revealed that TNFAIP3 and TNIP1 were key biomarkers for psoriasis. Here, we intended to conduct a survey on the association between TNFAIP3 and TNIP1 gene polymorphisms and psoriasis risk.A comprehensive search of four online databases—China National Knowledge Infrastructure (CNKI), PubMed, Embase, and Cochrane Library was undertaken up to August 25, 2019. We chose allele genetic model to deal with the original data. Newcastle–Ottawa scale (NOS) was used to evaluate the risk bias of each study. The RevMan 5.3 software was used to calculate the combined odds ratio and 95% confidence interval.P = 0.0002), and a significant association between rs17728338 in TNIP1 polymorphism and psoriasis vulnerability using fixed-effect model .In total, we included 13 case-control studies consist of 13,908 psoriasis patients and 20,051 controls in this work. Our results demonstrated that rs610604 in TNFAIP3 polymorphism was significantly associated with psoriasis risk using random-effect model (G vs. T, OR = 1.19, 95% CI: 1.09–1.31, Our findings indicated that rs610604 in TNFAIP3 and rs17728338 in TNIP1 gene polymorphisms were associated with psoriasis susceptibility. Psoriasis is currently regarded as a chronic, inflammatory skin disease associated with systemic conditions –3. As wiAs a complex inflammatory disorder, the aetiology and pathogenesis of psoriasis are widely thought to be caused by the interplay of intrinsic and environmental factors. Numerous triggers and aggravations for psoriasis occurrence have been identified such as mild localized trauma, drugs, HIV infection and streptococcal pharyngitis . HoweverTNFAIP3 gene is located on human chromosome 6q23.3, another alias for TNFAIP3 is A20. While TNIP1 is located on 5q33.1. They are all Protein-Coding genes encode ubiquitin-editing enzyme A20 and A20-Binding Inhibitor Of NF-Kappa-B Activation 1 (ABIN-1), which interact with each other to influence intracellular signaling , 11. PolTwo of our investigators (Hai-bo Gong and Shu-tao Gao) independently did the literature searching among four major databases—CNKI, PubMed, Embase, and Cochrane Library—for papers published before August 25, 2019. The retrieval strategy of PubMed was as follows: ((((Psoriasis) OR “Psoriasis”[Mesh])) AND  OR Tumor Necrosis Factor alpha-Induced Protein 3) OR Zinc Finger Protein A20) OR A20) OR TNF Alpha-Induced Protein 3)) OR  OR TNIP1) OR TNFAIP3 Interacting Protein 1) OR A20-Binding Inhibitor Of NF-Kappa-B Activation 1) OR VAN protein, human) OR TNFalpha-induced protein 3-interacting protein 1, human) OR ABIN-1 protein, human))) AND ((((((single nucleotide polymorphism) OR Polymorphism) OR Alleles) OR SNP) OR Variation) OR gene).The main contents of our inclusion criteria are given as follows: published case-control studies on humans; psoriasis should be diagnosed according to clinical diagnosis criteria; studies related to the association of TNFAIP3 or TNIP1 polymorphisms with psoriasis susceptibility; contained sufficient original data to estimate odds ratios (ORs) and 95% confidence intervals (95% CIs). Accordingly, Studies met the following criteria should be excluded: reviews, conference abstracts, case reports; the exact number of alleles wasn’t able to be ascertained, and duplicated articles.The following information was extracted from each candidate study by two independent investigators (Hai-bo Gong and Shu-tao Gao) including: author’s names, year of publication, Ethnicity of the study population, numbers of cases and controls, and the allele frequencies of the TNFAIP3 or TNAP1 polymorphisms, Hardy-Weinberg equilibrium (HWE) results.The methodological quality of each eligible study was evaluated by two investigators (Hai-bo Gong and Shu-tao Gao). Newcastle–Ottawa Scale (NOS) was used to appraise all included studies in line with its criteria. We used “Score system” to judge each study mainly made up of three aspects: selection (four items), comparability (two items), and the outcomes of case control studies (three items). Every included study received a total of 0–9 scores according to these items. A higher score indicated better quality. Studies with ≥5 scores were considered to have high quality for further analysis. When disagreements occurred between the two investigators, the third reviewer (Xiong-ming Pu) will be invited to make the final decision.I2 test were used to evaluate the heterogeneity among all included studies [P < 0.1, I2 > 50%) or fixed-effect model was used when it was out of heterogeneity  [The preferred Reporting Items for Systematic Review and Meta-analyses (PRISMA) was used to complete this meta-analysis . We proc studies . The ran2 < 50%) , 18. TheTo dig deeper into the potentially functional role of these loci, we used in silico tool HaploReg 4.1 to explore the annotations in the human genome.Initial retrieval of the four databases harvested 206 records: 43 from China National Knowledge Infrastructure (CNKI), 72 from PubMed, 80 from Embase, and 11 from Cochrane library. After removed duplicated and irrelevant records. 13 articles ultimately went into the process of meta-analysis. The detailed process of the literature search and screen is shown in Fig. P < 0.0001, I2 = 70%). The results demonstrated that rs610604 was significantly associated with psoriasis risk . therefore, the fixed-effect model was used to conduct the meta-analysis. Our results demonstrated a significant association between rs17728338 polymorphism and psoriasis vulnerability  and rs17728338 (P = 0.616) , which has another alias of ABIN-1. It has been reported that TNFAIP3 and TNIP1 physically interact with each other to inhibit cell death and NF-κB signalling pathway , 47. SimTo date, this is the first comprehensive study on the correlation between TNFAIP3 and TNIP1 polymorphisms and psoriasis vulnerability. However, several drawbacks should not be overlooked. First, as far as a limited number of studies are concerned, although we have collected all the currently related data, false negatives of our study may exist. Second, the genetic factor for psoriasis is composed of multiple genes and loci. However, we only concentrated on rs610604 in TNIP1 and rs17328338 in TNFAIP3 in this work. Third, only the allele model was used to analyze the data. Other genetic models are strongly recommended to be used as long as there are enough future relevant researches. Fourth, the HWE of some included studies were missing, which may lead to Information bias. Finally, only studies published in English were finally included in the present study, which may lead to selection bias.The results of our meta-analysis suggested that G allele of rs610604 polymorphisms in TNFAIP3 and A allele of rs17728338 polymorphisms in TNIP1 were considered to have an increased risk for psoriasis.


How credible is the evidence behind the association of dietary factors with colorectal cancer (CRC) risk in published meta-analyses of prospective observational studies?This umbrella review of 45 meta-analyses describing 109 associations found convincing evidence for an association between lower CRC risk and higher intakes of dietary fiber, dietary calcium, and yogurt and lower intakes of alcohol and red meat.This study suggests that dietary factors may have a role in the development and prevention of CRC, but more research is needed on specific foods for which the evidence remains suggestive. This umbrella review grades the evidence from published meta-analyses of prospective observational studies that assessed the association of dietary patterns, specific foods, food groups, beverages , macronutrients, and micronutrients with the incidence of colorectal cancer (CRC). Several meta-analyses have summarized evidence for the association between dietary factors and the incidence of colorectal cancer (CRC). However, to date, there has been little synthesis of the strength, precision, and quality of this evidence in aggregate.To grade the evidence from published meta-analyses of prospective observational studies that assessed the association of dietary patterns, specific foods, food groups, beverages , macronutrients, and micronutrients with the incidence of CRC.MEDLINE, Embase, and the Cochrane Library were searched from database inception to September 2019.Only meta-analyses of prospective observational studies with a cohort study design were eligible. Evidence of association was graded according to established criteria as follows: convincing, highly suggestive, suggestive, weak, or not significant.I2 > 50%), whereas small-study effects were found for 11 associations (10.1%). Excess significance bias was not detected for any association between diet and CRC. The primary analysis identified 5 (4.6%) convincing, 2 (1.8%) highly suggestive, 10 (9.2%) suggestive, and 18 (16.5%) weak associations between diet and CRC, while there was no evidence for 74 (67.9%) associations. There was convincing evidence of an association of intake of red meat (high vs low) and alcohol  with the incidence of CRC and an inverse association of higher vs lower intakes of dietary fiber, calcium, and yogurt with CRC risk. The evidence for convincing associations remained robust following sensitivity analyses.From 9954 publications, 222 full-text articles (2.2%) were evaluated for eligibility, and 45 meta-analyses (20.3%) that described 109 associations between dietary factors and CRC incidence were selected. Overall, 35 of the 109 associations (32.1%) were nominally statistically significant using random-effects meta-analysis models; 17 associations (15.6%) demonstrated large heterogeneity between studies (This umbrella review found convincing evidence of an association between lower CRC risk and higher intakes of dietary fiber, dietary calcium, and yogurt and lower intakes of alcohol and red meat. More research is needed on specific foods for which evidence remains suggestive, including other dairy products, whole grains, processed meat, and specific dietary patterns. The WCRF/AICR report applied many, but not all, umbrella review criteria for evidence grading and was based on studies published until April 2015.7 To complement the WCRF/AICR report, we conducted an umbrella review of meta-analyses to provide an updated and unified systematic summary of epidemiological data evaluating the strength of the overall body of evidence investigating the associations between dietary factors and CRC incidence. Furthermore, we expand the review of dietary exposures to some not previously evaluated, such as certain dietary patterns.Several systematic reviews with meta-analysis of prospective observational studies have summarized evidence for the associations between dietary factors  and the incidence of CRC. However, to date, there has been little synthesis of the strength, precision, and quality of this evidence in aggregate. Umbrella reviews provide a structured and critical summary of the evidence and enable the grading of evidence according to specific criteria: sample size, strength and precision of the association, and assessment of the presence of biases.CRD42020173636). This study adhered to the Meta-analysis of Observational Studies in Epidemiology (MOOSE) reporting guideline.The protocol of this umbrella review has been registered in PROSPERO  independently screened titles or abstracts and examined the full text of potentially eligible articles. Discrepancies were resolved by a third reviewer (S.K.V.).6 We excluded meta-analyses of studies with other study designs and those with insufficient or inadequate data for quantitative synthesis.Studies were included that met the following criteria: (1) meta-analysis of prospective observational studies  among adults with multivariable-adjusted summary risk estimates and corresponding 95% CIs that (2) investigated the association of dietary factor(s) with the incidence of CRC. Eligible dietary factors included dietary patterns, prespecified diet quality indices, specific foods, food groups, beverages , macronutrients , and micronutrients . When more than 1 meta-analysis on the same research question was available, we assessed only the study that included the largest data set, as previously described (eAppendix in the 8 by 2 authors (Y.S.L. and T.Y.W.) independently and checked by a third author (S.K.V.). For each eligible meta-analysis, we abstracted data at the meta-analysis level P values using the DerSimonian and Laird random-effects model.9 Heterogeneity was assessed with the I2 statistic.10 We estimated the 95% prediction interval (PrI), which evaluates the uncertainty for the effect size that would be anticipated in a new study addressing the identical association.11 The evidence for small-study effects was assessed by Egger regression asymmetry test.12P < .10 was taken as statistical evidence of the presence of small-study effects. We also applied the excess significance test, which evaluates whether the observed number of studies with statistically significant results  differs from the expected number of positive studies, by using a χ2 test.13 Excess significance bias was set at P ≤ .10. The statistical analysis and power calculations were conducted using Stata version 15.0 (StataCorp).For each association of diet with CRC, we recalculated the adjusted summary estimates and corresponding 95% CIs with 6 In brief, associations that presented nominally significant random-effects summary effect sizes  were graded as convincing (class I), highly suggestive (class II), suggestive (class III), or weak (class IV) evidence .For each meta-analysis initially graded as showing convincing, highly suggestive, or suggestive evidence, we reexamined the list of adjusted confounding factors at the component study level. We performed a sensitivity analysis by including only adjusted estimates to assess the robustness of the main analysis. Based on the literature, the most consistent potential confounders considered for sensitivity analysis are age, body mass index (BMI), race, sex, and other dietary factors as observed in our umbrella review. Other sensitivity analyses included the omission of small-sized studies 59 describing 109 associations in this umbrella review , and included 45 meta-analyses (20.3%)The evaluation of methodological quality using AMSTAR-2  revealed55 (13 associations), food groups59 (23 associations), beverages including alcohol33 (12 associations), macronutrients59 (18 associations), and micronutrients54 (43 associations) and the incidence of CRC. Definitions of dietary patterns is provided in the eAppendix in the Overall, the 45 meta-analyses described 109 associations, including 794 individual study estimates of CRC incidence associated with dietary exposures. The included meta-analyses provided adjusted summary estimates on the associations between dietary patternsP ≤ .05  suggested potential protective effects of dietary factors or dietary patterns associated with CRC risk, including adherence to a healthy diet, Mediterranean diet, pesco-vegetarian diet, or semivegetarian diet and higher intakes of dietary fiber, whole grains, legumes, dairy products including yogurt and nonfermented milk, fruits and vegetables, and micronutrients . The remaining significant associations (31.4%) suggested higher risk of CRC with adherence to an unhealthy diet or Western diet and increased intake of alcohol, red meat, processed meat, pork, eggs, and haem iron.A total of 35 of the 109 associations (32.1%) were nominally statistically significant at P ≤ .05 . Of thesI2 > 50%). The 95% PrIs excluded the null value for 8 associations (7.3%). The effect sizes of the largest study were statistically significant at P ≤ .05 for 29 associations (26.6%). Small-study effects were found for 11 associations (10.1%), and excess significance bias was not identified. Seventy-four (67.9%) associations without statistical significance at P ≤ .05 using random-effects models are presented in eTable 4 in the Seventeen associations (15.6%) had large heterogeneity  were supported by convincing evidence . Two of Two associations (1.8%) had highly suggestive evidence of an association between diet and incidence of CRC . Higher Suggestive evidence for the association of diet or dietary patterns with reduced risk of CRC was found for 8 associations (7.3%), including adherence to the Mediterranean diet, adherence to a healthy diet, adherence to a pesco-vegetarian diet, adherence to a semivegetarian diet, and intake of whole grains, nonfermented milk, and supplemental calcium . HoweverResults from sensitivity analyses are reported in eTable 5 in the We included 45 published meta-analyses, which comprised 109 adjusted summary risk estimates for the associations of dietary factors with CRC incidence. We found that few of the 35 statistically significant associations  were supported by convincing evidence in the main and sensitivity analyses. Suggestive evidence exists for an inverse association of a healthy dietary pattern, Mediterranean diet, pesco-vegetarian diet, and semivegetarian diet and intake of whole grains, total dairy products, and supplemental calcium with CRC incidence. Suggestive evidence also exists for positive associations between higher intakes of processed meat and a moderate intake of alcohol (>1-3 drinks/d) and the incidence of CRC.60 Although the WCRF/AICR report provides rigorous estimates for the association of diet with CRC incidence, their criteria for study inclusion and grading of evidence differed from the current umbrella review . In addition, yogurt, which was not evaluated as a discrete exposure in the WCRF/AICR CUP report, was identified as being associated with reduced risk of CRC.The continuous update project (CUP) report by the WCRF in partnership with the AICR conducted a comprehensive review including meta-analyses to draw conclusions about the roles of diet, nutrition, and physical activity in cancer prevention and survival.63 However, we did not find convincing evidence to support limiting consumption of processed meat for CRC prevention. Our results confirmed the positive association of processed meat products with CRC; however, its credibility was suggestive. We did not measure dose-response associations owing to data limitations. However, we examined the summary effect size estimate for concordance regarding the direction of association and level of statistical significance to other eligible meta-analyses reporting the same association . We observed a weak inverse association between fruit and vegetable consumption and CRC incidence. Fruit and vegetables are abundant sources not only of fiber, but also of micro- and macronutrients with antitumor properties; thus, they are plausible targets for dietary prevention.66 Whether dietary calcium is a causal factor in CRC prevention is difficult to determine given that other correlated constituents  could account for the observations.67 The high fat content of cheese and cream may hinder their protective associations, possibly by increasing bile acid and fatty acid excretion in the colonic lumen. Other nutrients or bioactive compounds in dairy products, such as lactoferrin or generation of butyrate, may also play a role.67 The biologic plausibility of yogurt’s associations with reduced risk of CRC stems from the presence of lactic acid–producing bacteria (Lactobacillus bulgaricus and Streptococcus thermophilus) that are purported to reduce levels of carcinogens such as nitroreductase, fecal-activated bacterial enzymes, and soluble fecal bile acids.70 Yogurt intake has been shown to reduce the risk of colorectal adenomas with high malignant potential independent of calcium and nonyogurt dairy intake.71 Its protective association could also be mediated through its calcium content and modulated by the gut microbiome.Our findings convincingly highlight dietary calcium and yogurt as associated with reduced risk of CRC incidence. Findings for cheese and milk were null and weak, respectively. We also found suggestive evidence for lower CRC risk with higher intakes of supplemental calcium. These findings are complementary given that all dietary factors are sources of calcium, which binds to unconjugated bile acids and free fatty acids in the colonic lumen to minimize their toxic effects.Selected meta-analyses provide suggestive evidence that supports the adoption of an overall healthy dietary pattern, Mediterranean diet, pesco-vegetarian diet, and semivegetarian diet to prevent CRC, in contrast with Western dietary patterns that are associated with increased CRC risk. Overall dietary pattern accounts for the totality of dietary components in combination and their synergistic or antagonistic influence on human metabolism and disease. Prudent dietary patterns, characterized by higher intakes of vegetables, fruit, whole grains, and low-fat dairy products and lower intakes of alcohol and meat products, are often accompanied by a healthier lifestyle; the converse is true for Western-type dietary patterns. Hence, the suggestive association observed for these dietary behaviors could be influenced by the totality of a healthy or unhealthy lifestyle, in addition to combination(s) of protective or harmful dietary factors.72 Although the small-study effect was indicated for only 11 meta-analyses (10.1%), among these 4 (36.4%) included 5 to 10 studies and 2 (18.2%) included less than 5 studies. However, we were unable to perform alternate tests, such as the Peters test, because this required cases and noncases for each level of exposure, and this information was largely lacking in the meta-analyses.73 Thus, more research is needed to investigate the associations based on small numbers of included studies. A further limitation is that we did not conduct subgroup analysis  because of the lack of data for grading the quality of evidence for most of the exposures. Dietary associations with CRC may differ according to sex and tumor location, as was highlighted by the WCRF/AICR reports.60This study has limitations. A possible limitation of our review was exclusion of dose-response meta-analyses because the data needed for predictive interval estimation and assessment of small study and excess significant bias effects were not available in those articles. Randomized clinical trials, which account for confounding by design, are scarce in research on associations between diet and cancer owing to cost, the long follow-up time required for cancer end points, and ethical concerns. Consequently, we restricted the current umbrella review to meta-analyses of prospective observational studies. Some forms of bias, such as recall bias owing to self-reported diet, are possible but likely to be nondifferential, which would attenuate the observed associations. We generated estimates of publication bias by assessing small-study effects using the Egger test. However, the Egger test is not recommended with the inclusion of less than 10 studies.The findings of this study support existing recommendations for diet in the primary prevention of CRC, emphasizing higher intakes of dietary fiber, calcium, and yogurt and lower intakes of red meat and alcohol. Emerging evidence supports a possible role for overall dietary patterns that, in totality, emphasize habitually consuming fruits, vegetables, grains, and low-fat dairy and reducing red meat and alcohol intake. More research is needed on specific foods for which evidence remains suggestive, including other dairy products, whole grains, processed meat, and specific dietary patterns.