I2 = 37.6%). On the other hand, 9 studies with data on 46,487 participants analyzed the effects of NAFLD on incident hypertension. Pooled analysis showed that the presence of NAFLD was significantly associated with an increased incidence of hypertension . There was significant heterogeneity among the studies in this analysis (p < 0.01). Sensitivity analyses showed that the magnitude of the association was significantly different in subgroups stratified by a mean age of participants and geographical location, which explains part of the heterogeneity. In conclusion, this meta-analysis indicates the existence of a bidirectional relationship between NAFLD and hypertension independent of traditional cardiometabolic risk factors.An increasing body of evidence connects non-alcoholic fatty liver disease (NAFLD) to hypertension. The objective of this systematic review and meta-analysis was to estimate the nature and magnitude of the association between NAFLD and hypertension. We systematically searched PubMed, Embase, Cochrane Library, and Web of Science for observational studies published up to May 1, 2021. Cohort studies that reported data on the association between NAFLD and incident hypertension or between hypertension and incident NAFLD were included. We used random-effects models to conduct meta-analysis on the measures of association from individual studies. A total of 11 studies were eligible for inclusion, among which 4 studies including 25,260 participants reported the association between hypertension and new-onset NAFLD. The presence of hypertension was significantly associated with an increased risk of incident NAFLD (HR 1.63, 95% CI: 1.41–1.88;  Nonalcoholic fatty liver disease (NAFLD) encompasses a pathological spectrum ranging from simple steatosis through steatohepatitis (NASH) to advanced fibrosis, cirrhosis, and ultimately hepatocellular carcinoma , 2. NAFLHypertension (HTN), a major component of metabolic syndrome affecting around 30% of the general population, has long been recognized as one of the strongest risk factors for CVD , 7. EmerGiven the huge number of patients with NALFD and HTN worldwide, elucidating their relationship is important for risk prediction, primary prevention, and exploring therapeutic regimens for these two diseases. We therefore attempted a synthesis of the available evidence from observational cohort studies under a meta-analysis to examine the relationship between NAFLD and HTN and to assess the magnitude of this association.We did a systematic literature search of PubMed, Embase, Cochrane Library, and Web of Science for the identification of articles reporting the results of longitudinal observational studies published up to May 1, 2021, investigating the associations of NAFLD on incident HTN and/or of HTN on incident NAFLD. We used a combined text and medical subject headings (MeSH) search strategy with the terms: “Nonalcoholic Fatty Liver Disease,” “NAFLD,” “Non alcoholic Fatty Liver Disease,” “Nonalcoholic Fatty Liver,” “Nonalcoholic Fatty Livers,” “Liver, Nonalcoholic Fatty,” “Livers, Nonalcoholic Fatty,” “Fatty Liver, Nonalcoholic,” “Fatty Livers, Nonalcoholic,” “Nonalcoholic Steatohepatitis,” “Nonalcoholic Steatohepatitides,” “Steatohepatitis, Nonalcoholic,” “Steatohepatitides, Nonalcoholic,” “Hypertension,” “High Blood Pressure,” “High Blood Pressures,” “Blood Pressure, High,” and “Blood Pressures, High.” The search was restricted to studies in human beings, and no language restrictions were applied. Full electronic search strategies for PubMed and Embase are shown in Studies that met the following criteria were included: (1) longitudinal design; (2) participants aged ≥18 years; (3) follow-up duration ≥2 years; (4) assessment of the association between NAFLD and incident HTN and/or between HTN and incident NAFLD; (5) reported data on hazard ratios (HRs) or odds ratios (ORs) with confidence intervals (CIs) for the outcome of interest, and the estimate was at least adjusted for age and gender; and (6) NAFLD diagnosed by aminotransferase levels, surrogate scores, imaging methods, or liver biopsy and HTN diagnosed based on blood pressure. Criteria for exclusion of selected articles were as follows: (1) cross-sectional studies, (2) studies that were conducted in the pediatric population, (3) studies with a median follow-up less than 2 years, (4) studies did not report the adjusted point estimate, and (5) review articles, case reports, practice guidelines, commentaries or editorials. Two authors (GL and YP) independently assessed all titles and abstracts and reviewed full texts of potentially relevant articles for inclusion. Any disagreement was resolved by consensus, referring back to the original paper.All data were extracted into a prepiloted data extraction form developed in Excel. For all included studies, we extracted information on study design, country/region, follow-up duration, diagnostic methods of NAFLD and HTN, the outcome of interest, risk estimates (95% CI), and adjustment factors. The most adjusted estimate in the multivariable regression model was extracted when a study presented more than one risk estimate. In the case of multiple articles on the same subjects, we included only the latest and most comprehensive one. Two authors (GL and YP) conducted quality evaluation independently, and any disagreement in rating was addressed by discussion. The Newcastle–Ottawa Scale (NOS) was applied for the quality evaluation of included cohort studies. The quality of each study ranges from 1 to 9 points based on three domains: the cohort selection (maximum of 4 points), the comparability of the cohort design and analysis (maximum of 2 points), and the adequacy of the outcome measures (maximum of 3 points). Studies achieving a score of at least seven points were considered high quality .I-squared (I2) statistic were applied to assess the heterogeneity across studies. Large I2 (>50%) or p < 0.1 for Q-statistic suggests substantial heterogeneity across studies.For assessing the effect of NAFLD on the risk for incident HTN, the effect size was estimated as HRs or ORs with 95% CIs, according to NAFLD patients versus non-NAFLD patients shown in each included study, using non-NAFLD patients as the reference group. For evaluating the effect of HTN on the risk for incident NAFLD, the effect size was estimated as HRs or ORs with 95% CIs, according to HTN patients versus non-HTN patients reported in each eligible study, using non-HTN patients as the reference. The results were pooled, and an overall estimate of effect size was calculated using a random-effects model to take into account the heterogeneity among studies, as we expected a relatively large heterogeneity in our pooled results of observational studies on different cohorts with varying degrees of adjustment. The Q-statistic and the p-value <0.05 was considered significant.A funnel plot showing the logarithm of the effect measure against the logarithm of its standard error was constructed to evaluate the presence of publication bias. We also applied both the Egger's test  and the As specified in the PRISMA flow diagram , our iniThe main characteristics of the 11 included studies are summarized in n = 4 studies, 25,260 participants). The test for heterogeneity among the individual studies was not significant  and rank correlation Begg's test (p = 0.734) did not show statistically significant asymmetry . We conducted subgroup analyses to explore potential sources of heterogeneity (n = 5 studies), a higher risk was seen in studies where NAFLD was diagnosed by surrogate scores . However, no significant results were found in the test for differences between subgroups stratified by NAFLD diagnostic methods (p = 0.17) as well as the follow-up duration (p = 0.43). In contrast, a significant between-group heterogeneity was found in subgroup analyses stratified by mean age of participants (p = 0.02) and geographical region (p = 0.03). Asymmetry analysis in the funnel plot showed no evidence of significant publication bias, and both Egger's test (p = 0.062) and rank correlation Begg's test (p = 0.118) did not show statistically significant asymmetry . On the other hand, the presence of HTN was significantly associated with a higher incidence of NAFLD . The results from longitudinal cohort studies clearly suggest a bidirectional relationship between NAFLD and HTN, and this association appears to be independent of traditional cardiometabolic risk factors. The close two-way association between the two entities can conceivably form a vicious circle during disease progression. It is therefore important to increase public awareness of the notion that NAFLD is an independent risk factor and might serve as a driving force in the development and progression of HTN and vice versa.In this meta-analysis of 11 observational cohort studies, we found that the presence of NAFLD is significantly associated with a higher risk of incident HTN . Compared with the meta-analysis by Wu et al., we have increased the number of eligible studies to 9 in the analysis of the association between NAFLD and incident hypertension. Moreover, we further analyzed the association between hypertension and incident NAFLD, thus enabling the investigation of a possible bidirectional relationship.The findings of this meta-analysis confirm and extend the results of a previous smaller meta-analysis published in 2016 . In thatIn light of our results and given the extremely high prevalence of both NAFLD and HTN in the general population, identifying NAFLD as an emerging risk factor for HTN and vice versa may help improve the risk prediction, identify primary preventive strategies, and select therapeutic regimens for the two diseases and their related complications , 19. IndDespite the growing body of clinical evidence linking NAFLD to HTN, the mechanisms underpinning this association has yet to be identified  5–7]. I–7. I5–7]p = 0.17). Moreover, covariate adjustment and potential unmeasured variables can serve as another important source of heterogeneity. As shown in The present meta-analysis has several limitations. First, the observational nature of the included studies could not allow the interpretation of a causality link between the exposure and the outcome. Second, the study protocol of this systematic review and meta-analysis has not been preregistered. Third, in the present study, the pooled analysis of the association between NAFLD and incident HTN revealed a certain level of heterogeneity. Thus, these results should be interpreted cautiously. The heterogeneity may be derived from a combination of factors including diagnostic methods for NAFLD, age of participants, geographical region, covariate adjustment, and other potential unmeasured confounders. The included studies have defined NAFLD by various diagnostic methods, resulting in variations in their reported outcomes. Our subgroup analysis stratified by NAFLD diagnostic methods showed that compared with the studies where NAFLD was diagnosed by imaging methods (HR 1.38), a higher risk was seen in studies where NAFLD was diagnosed by surrogate scores (HR 1.81), although no significant results were found in the test for differences between the subgroups (Albeit these limitations, our meta-analysis also has several important strengths. To our knowledge, this is the first meta-analysis to directly and specifically analyze the bidirectional association between NAFLD and HTN. In addition, we conducted a comprehensive search, used predetermined inclusion and exclusion criteria, and did a robust quality assessment, which means we have included the best available evidence to report on the association between the two diseases. Moreover, the large number of participants and events involved in our meta-analysis yields adequate statistical power to quantify the association between NAFLD and HTN. Finally, the overall quality of the included studies in the meta-analysis was considered as high, and there was no sign of significant publication bias affecting the results when evaluated by both Egger and Begg's tests.In conclusion, our meta-analysis clearly suggests the existence of a bidirectional relationship between NAFLD and HTN independent of traditional cardiometabolic risk factors. The presence of NAFLD is significantly associated with a 1.55-fold increased risk of incident HTN, whereas HTN is significantly associated with a 1.63-fold increased risk of incident NAFLD. Recognizing NAFLD as a risk factor for HTN and vice versa is important to improve the risk prediction, identify primary preventive strategies, and explore therapeutic regimens for these two increasingly prevalent and burdensome diseases. More studies in the future should investigate whether changes in NAFLD status over time can modify the risk of developing or worsening HTN and vice versa, thus providing more solid evidence to definitively establish a cause-effect relationship. In addition, further researches are needed to gain an in-depth understanding of the underlying mechanisms and to identify novel and effective treatments.


Background: There is increasing evidence that nephrolithiasis is a systemic disease, as opposed to an isolated urinary metabolic problem, after considerable links were found between nephrolithiasis and systemic diseases such as hypertension, obesity, dyslipidemia, and insulin resistance. The interplay between these four factors defines metabolic syndrome (MetS). In this review we aim to clarify the associations of MetS and its components to kidney stone incident.Methods: Online databases of EMBASE, MEDLINE, and Google Scholar were searched from January 1998 up to October 2020 to identify observational studies examining the association between metabolic syndrome components and kidney stone incident. Bayesian random-effects meta-analysis and meta-regression were performed to observe the association. Linear dose-response analysis was conducted to shape the direction of the association. Data analysis was performed using STATA, and R statistics.Results: A total of 25 potentially relevant studies  were eventually identified. The pooled results suggested that metabolic syndrome was associated with an increased risk of nephrolithiasis with an odds ratio (OR) of 1.769 (95% CI: 1.386 – 2.309).  The summary OR of hypertension and dyslipidemia for developing nephrolithiasis were 1.613 (95% CI: 1.213 – 2.169) and 1.586 (95% CI: 1.007 – 2.502) respectively. The presence of diabetes mellitus and obesity had an OR of 1.552 (95% CI: 1.027 – 2.344) and 1.531 (95% CI: 1.099 – 2.109) respectively. Our results revealed that the increasing number of MetS traits will increase the risk of developing nephrolithiasis, the higher the fasting plasma glucose, and body mass index, the higher the risk of kidney stones incident.Conclusions: Our results suggest that hypertension, diabetes, obesity and dyslipidemia are associated with increased risk of developing nephrolithiasis. Linear significant association between MetS components and nephrolithiasis were revealed in our study which reinforced the notion that should be considered a systemic disorder. Kidney stone disorder, characterized by abnormal urine composition and dehydration, is an increasingly common condition that is most common in older men  using specific keywords combined with medical subject heading (MeSH) terms and words: "metabolic syndrome ", "hypertension", "diabetes mellitus", "insulin resistance", "dyslipidemia", "obesity", “waist circumferences”, “hypertriacylglycerolemia”, “hypertriglyceridemia”, “kidney stones”, “nephrolithiasis". The full texts and abstracts were originally and independently selected by two reviewers. Full texts and abstracts were scored according to the inclusion and exclusion criteria. The studies which did not fit with the inclusion criteria were excluded. Discrepancies between the two reviewers were resolved through a discussion with a third reviewer. Ethical approval was not required because the data did not contain individual patient information.The studies included in this article met the following criteria: (1) the subjects were nephrolithiasis patients; (2) research was conducted as observational study to show causal-effect relationship (case control or cohort studies); (3) the study showed association as risk with 95% CI; (4) the language of the article was published in English. Two reviewers (IAR and IF) separately extracted and analyzed the data based on study selection criteria using standardized, structured and pilot extraction forms. The results were reviewed and discussed by IAR and IF to complete the included studies. Any discrepancies were resolved in a discussion with a third reviewer. For each included study, several pieces of information were extracted including author name, year of publication, number of sample sizes, mean age, study design, countries, number of participants and cases, study period, predictors and outcomes.Primary endpoint was the occurrence of kidney stones with the predictors assessed as metabolic syndrome traits. Secondary endpoint was the direction of association assessing between the dose of the predictors and the response of kidney stones occurrence.2 statistic quantification were used to assess heterogeneity of studies. I2 values of heterogeneity were interpreted as insignificant (0–25%), low (26–50%), moderate (51–75%) and high (>75%) , STATA  and R. An odds ratio (OR) with a 95% confidence interval (CI) used for each association was generated using Bayesian meta-analysis. The comparison of risk factors with each other was conducted based on OR, where HRs (Hazard Ratio) and RRs (Risk Ratio) were both directly considered as OR , for kidThis meta-analysis study included a total of 207 studies found using the search strategy. A full text review and abstract selection were then performed and 138 studies were excluded due to reviews, case reports, duplicates, and not in English. Finally, 25 studies met the inclusion criteria of 934,588 patients, which were eligible and added for further analysis. A block diagram of the procedure for searching and selecting studies is shown inAs demonstrated inAs demonstrated in2 = 97%)The results of meta-analyses that assessed the relationship between diabetes and nephrolithiasis were shown in2 = 96%) between studies, the results of sensitivity analyses remained consistent in different values of study variances  and Egger’s (P = 0.5712) tests  were found to be positively correlated , trim-fill (b) and sensitivity analysis (c) of the association between metabolic syndrome and nephrolithiasis-Supplementary material 2. Funnel plot (a), trim-fill (b) and sensitivity analysis (c) of the association between hypertension and nephrolithiasis-Supplementary material 3. Funnel plot (a), trim-fill (b) and sensitivity analysis (c) of the association between DM and nephrolithiasis-Supplementary material 4. Funnel plot (a), trim-fill (b) and sensitivity analysis (c) of the association between obesity and nephrolithiasis-Supplementary material 5. Funnel plot (a), trim-fill (b) and sensitivity analysis (c) of the association between dyslipidemia and nephrolithiasisCreative Commons Zero "No rights reserved" data waiver (CC0 1.0 Public domain dedication).Data are available under the terms of thehttps://doi.org/10.6084/m9.figshare.13585607 .Data are available under the terms of the This study assessed the association between metabolic syndrome and its components with the risk of developing nephrolithiasis by conducting systematic review, Bayesian random-effects meta-analysis, meta-regression and dose-response analysis. This study was done appropriately based on PRISMA flowchart. Risk of bias was also conducted of the included studies. This study has successfully presented the proper meta-analysis for this design. However, to complete this study for indexing, I personally recommended several revisions:Introduction section: It is better to address meta-regression as the analysis to assess the correlation of association along with dose-response analysisConclusion section: In reporting the association between predictors and nephrolithiasis, state only the predictors in which its coefficient was statistically significant 2. R language was not considered as a statistical software for data analysis. The software for data analysis should be written as "R" . 1. Abstract: 3. Please update the PRISMA flowchart (refer to PRISMA guideline 2009). 4. Give the numbering of each Forrest plot in Figure 3 and numbering of each meta-regression plot in Figure 4. Design these figures so that it could be well presented. 5. Uniformly decide the word choice of "traits" or "components", choose whether to use traits or components in the whole text, use one of these words consistently to avoid any misunderstanding. 6. It is better to provide the meta-regression of hypertension in systolic blood pressure and diastolic blood pressure as it is important to explain the relationship to nephrolithiasis in differentiation for these two types of blood pressure. 7. Meta-regression of body mass index  was sufficient in this study thus waist circumference meta-regression was not necessary to be included. 8. Provide the value of coefficient and confidence interval of each meta-regression analysis in the result section so that better understanding of predictors-outcome relationship could be reached clearly.Are the rationale for, and objectives of, the Systematic Review clearly stated?YesIs the statistical analysis and its interpretation appropriate?YesAre sufficient details of the methods and analysis provided to allow replication by others?YesAre the conclusions drawn adequately supported by the results presented in the review?YesReviewer Expertise:General Surgery, UrologyI confirm that I have read this submission and believe that I have an appropriate level of expertise to confirm that it is of an acceptable scientific standard. This systematic review is appropriate for the journal with a global problem of Mets and Urolithiasis. The introduction part clearly explains the motivation. The manuscript is clear and balanced. The manuscript stays focused on the subject. Authors have gone through the process of searching relevant articles from all websites and of sufficient duration. The inclusion and exclusion criteria in the analysis have been clearly stated. The impact of the analysis is clearly stated. The statistical analysis supports the paper well. The interpretation of the results, visualisation are well presented. The tables and figures are clear, relevant and correct. The authors demonstrate the knowledge of basic composition skills, including word choice, sentence structure, paragraph development and grammar.Limitations:             The studies included in the meta-analysis have cross-sectional nature and hence ascertainment of temporal association is not possible which also dictates need for further prospective studies. The specific type of stone formation is not correlated with studies. Despite these limitations all studies included in the meta-analysis showed the same directionality in the association between urolithiasis and Mets.Are the rationale for, and objectives of, the Systematic Review clearly stated?YesIs the statistical analysis and its interpretation appropriate?YesAre sufficient details of the methods and analysis provided to allow replication by others?YesAre the conclusions drawn adequately supported by the results presented in the review?YesReviewer Expertise:Endourogy, Urolithiasis, Andrology and Male Infertility, Uro-OncologyWe confirm that we have read this submission and believe that we have an appropriate level of expertise to confirm that it is of an acceptable scientific standard.


The aim of this study was to quantitatively summarize the available evidence on the association of breastfeeding with the risk of childhood cancer.A literature search of PubMed and Embase databases was performed to identify eligible observational studies published from inception to July 17, 2020. The categorical and dose-response meta-analysis was conducted by pooling relative risk (RR) or odds ratio (OR) estimates with 95% confidence intervals (CIs). Potential sources of heterogeneity were detected by meta-regression and stratification analysis. Sensitivity analysis and publication bias test were also carried out.P non-linear < 0.001). The most protective effect  was observed at a breastfeeding duration of 9.6 months. Four studies examined, the association between breastfeeding and risk of childhood neuroblastoma, and significant inverse associations were consistently observed in both the comparisons of ever breastfeeding versus non/occasional breastfeeding  and longest versus shortest breastfeeding . However, no associations of breastfeeding with risk of other cancers were found.Forty-five articles involving 475,579 individuals were included in the meta-analysis. Among the thirty-three studies on the association between breastfeeding and risk of childhood leukemia, the pooled risk estimates were 0.77  and 0.77 (95% CI 0.63–0.94) for ever versus non/occasional breastfeeding and longest versus shortest breastfeeding duration group, respectively. There was clear indication for non-linear dose-response relationship between breastfeeding duration and the risk of childhood leukemia (Our study supports a protective role of breastfeeding on the risk of childhood leukemia, also suggesting a non-linear dose-response relationship. Further studies are warranted to confirm the association between breastfeeding and risk of childhood neuroblastoma.The online version contains supplementary material available at 10.1186/s12916-021-01950-5. Childhood cancer is emerging as a major cause of death in children worldwide, which is bleaker for children with cancer in lower-middle-income countries . AlthougIt is widely accepted that breastfeeding may protect mothers against breast cancer  and ovarA previous meta-analysis of the association between breastfeeding and the risk of childhood cancer was based on 26 original studies published up to June 2004 . The pooThe study was registered in the international prospective register of systematic reviews (PROSPERO: CRD42020199446). The Preferred Reporting Items for Systematic reviews and Meta-Analysis (PRISMA) checklist for reporting the meta-analysis was shown in Additional file Original articles from PubMed and EMBASE databases were systematically searched from the inception to 17 July 2020 to identify potentially eligible studies on the association between breastfeeding and the risk of childhood cancer. The search strategy was as follows: (“child” OR “pediatric” OR “childhood” OR “children”) AND  AND (“breastfeeding” OR “infant feeding” OR “infant nutrition”). In addition, we conducted manual retrieval of the relevant references.The eligible studies were included as follows: (1) study design of cohort or case-control, (2) clearly defining the outcome of interest as cancer of specific anatomical site –23, (3) We extracted crucial information from the final studies retained, including first author, year of publication, data collection years, country or region, sample size, age in years, source of participants, cancer site, method of assessing breastfeeding, breastfeeding category, variables adjusted or matched, and corresponding risk estimates with 95% CIs.Two researchers (QY and PZ) independently rated the quality of the included studies using the Newcastle-Ottawa Scale with scores ranging from 0 to 9 points . This scAll statistical analyses were performed using STATA version 14.0. The multivariate-adjusted risk estimates were selected if they were reported in the original article; otherwise, the unadjusted risk estimates were calculated using the original data. For the studies reported risk estimates relative to a reference category other than shortest breastfeeding duration, the risk estimates were recalculated using the shortest breastfeeding duration as reference by using the Orsini method . When poIn the analysis of breastfeeding versus non/occasional breastfeeding, if the corresponding estimate had not been presented in a study, estimates associated with different breastfeeding categories were synthesized into a single estimate. We also combined the risk estimates comparing the longest with the shortest breastfeeding duration among the studies with equal or greater than three different breastfeeding categories. The forest plot of the association between breastfeeding and the risk of childhood cancer was generated for breastfeeding and non/occasional breastfeeding and longest versus shortest breastfeeding duration, respectively.A two-stage dose-response meta-analysis  was condI-squared (I2) [The heterogeneity was evaluated by Q-statistic test and red (I2) , 29. Thered (I2) . SubgrouFunnel plot was generated, and the symmetry means no potential publication bias. Publication bias was assessed using Egger’s test  and BeggThe flowchart of study selection is presented in Additional file 1: Figure S1. The primary search strategy for PubMed and Embase yielded 2905 and 2771 articles, and manual search from the reference lists of original studies or relevant reviews and meta-analyses on this topic yielded 67 additional articles. After removal of duplicates, 5545 articles were retrieved for assessment based on title and abstract, of which 116 articles were included for full text evaluation. After exclusion of 8 articles with insufficient data, 4 articles only reporting data on all cancers, 47 articles as meta, review, comment or meeting abstract, 2 articles conducted not among children, and 10 articles with overlapping data sets , a total of 45 articles with 475,579 participants were included for this meta-analysis , 33–73.n = 20), Asia (n = 8), and North America or Oceania (n = 17). There were 32 articles with population-based case-control design, 12 articles with hospital-based case-control design, and one article with cohort design. The median quality score of all included articles was 7, which resulted in 33 articles with a score of 7 or more and 12 articles with a score less than 7. Based on the ROBINS-I tool, 37 studies were considered at moderate risk of bias, and 8 studies were rated at serious risk of bias. Of the 45 articles, 33 studies provided the effect estimates for the association of breastfeeding on leukemia, 11 studies on lymphoma, 7 studies on brain tumors, 4 studies on neuroblastoma, 4 studies on soft-tissue sarcoma, 3 studies on nephroblastoma, 2 studies on retinoblastoma, and 2 studies on germ cell tumors. The details of included studies for the subsequent subgroup analysis are shown in Additional file 1: Table S3. Moreover, after excluding studies with the breastfeeding only as dichotomous variable, and no sufficient data of the number of cases and controls in each breastfeeding category, 23 studies were included for leukemia [The characteristics of selected studies are shown in Table leukemia , 70–72, leukemia , 43, 46,leukemia , 64, 67 P < 0.001 for non-linearity; Fig. P value of 0.075 by Begg’s test and 0.173 by Egger’s test.The summary risk estimates of the risk of childhood leukemia were 0.77  for breastfeeding versus non/occasional breastfeeding and 0.77  for longest versus shortest breastfeeding duration  and definition of reference category (P in meta-regression = 0.025) , but not associated with acute myeloid leukemia (AML) risk  and 0.77  for the comparison of breastfeeding versus non/occasional breastfeeding and longest versus shortest breastfeeding duration, respectively  and definition of reference group (P = 0.044) was identified by meta-regression analysis . There was no sign of asymmetry with a P value of 0.764 by Begg’s test and 0.261 by Egger’s test  and age (37/45), and most studies also matched participants using geographic location (15/45), and a few additionally used race or ethnicity (6/45). These matching variables are crucial for the comparability of cases and controls. Moreover, parental socioeconomic status (SES) was commonly used as adjustment, because higher SES among controls may overestimate the protective effect of breastfeeding on the risk of childhood cancer . SmokingCase-control study is the main study design that all included studies of the meta-analysis except one were case-control studies. Since most studies have collected exposure information through parental interview, case-control studies are susceptible to recall bias and selection bias. Cohort studies are considered to provide more robust estimates than case-control studies; therefore, further cohort studies are needed to provide more evidence on the association and risk of childhood cancer. An optimal study might be conducted within the framework of a large population-based registry or cohort with full access to baseline information regarding demographic characteristics, detailed data of breastfeeding including breastfeeding duration, the use of infant formulas, the main types of milk given, the age of introduction of a range of foods and so on, and collection of medical records to accurately identify all diagnosed cases . HoweverThere is sufficient evidence to show that breastfeeding was inversely associated with the risk by pooling a number of original articles, with 23% lower risk of childhood leukemia (95% CI 9–35%) for breastfeeding versus non/occasional breastfeeding and 23% lower risk (95% CI 6–37%) for longest versus shortest breastfeeding duration, respectively. Moreover, we found a protective effect of breastfeeding on the risk of childhood ALL among the studies with higher quality and population-based case-control studies.We found high heterogeneity in the meta-analysis on the association between breastfeeding and the risk of childhood leukemia, and there exists significant heterogeneity in different regional groups by meta-regression analysis. In the subgroup analysis of geographic location, we found that breastfeeding was more strongly associated with the risk of childhood leukemia in Asia. In terms of risk of bias, Robins-I tool rated 4 European studies and 2 Asian studies at serious risk, and 10 European studies and 6 Asian studies were considered at moderate risk. At the original study level, one aspect that deserves particular attention is the difference in the number of cases and the source of population, which may at least in part explain the heterogeneity observed across the geographic region. For example, the Asian subgroup analyses included 8 studies with a total of 1739 cases, with only 1 study enrolled more than 200 cases. Moreover, few studies (3/8) used population-based case-control study design. However, European articles included 14 studies with 7518 cases and 71% studies used population controls. This is a similar situation to North America or Oceania, in that a majority (6/11) of articles included in the meta-analysis had more than 200 case numbers, and most (9/11) used population controls. Therefore, the differences in risk estimates could be related to study quality issues. On the other hand, there were several other potential explanations could be proposed. A great variation in breastfeeding duration across the countries may result in the heterogeneity. It should be noted that breastfeeding duration is shorter in high-income countries than in those that are resource-poor. It was estimated that 25% of infants in the USA and Europe are exclusively breastfed through 6 months , 84, as Additionally, we performed the dose-response meta-analysis, which showed a specific non-linear dose-response relationship between breastfeeding and the risk of childhood leukemia. World Health Organization (WHO) and United Nations International Children’s Emergency Fund (UNICEF) developed the global strategy for infant and young child feeding that infants should be exclusively breastfed for the first 6 months of life to achieve optimal growth, development and health . In the There were several potential explanations why breastfeeding may decrease the risk of childhood leukemia. Breast milk contains high levels of immunologically active components and multifactorial anti-inflammatory defense mechanisms that influence the development of the immune system of the breastfed infant , 90. ForThe meta-analysis provided no convincing evidence of the association between breastfeeding and risk of lymphoma. The findings of the present study are consistent with previous observational studies although some early studies indicated negative association of breastfeeding with lymphoma risk, especially in Hodgkin’s lymphoma , 43. In Our pooled analysis presented an inverse association between breastfeeding and risk of childhood lymphoma in Asia, but not Europe or North America, suggesting a potential region-specific effect. What is more, significant discrepancies across subgroups stratified by definition of reference category were found, and only the pooled estimate for the studies using occasional breastfeeding as reference group showed a significantly decreased risk of childhood lymphoma. It may be explained by the significant non-linear dose-response relationship. For example, Davis et al.  found thWe found significant association between breastfeeding and risk of childhood neuroblastoma, while no significant associations of breastfeeding with risk of soft-tissue sarcoma, brain tumors, nephroblastoma, retinoblastoma, and germ cell tumors. This is updated from the previous meta-analysis and the results were consistent . HoweverThe primary strength of our study is that the traditional categorical meta-analysis and dose-response analysis were applied simultaneously, which can provide more meaningful information. Another strength is the large sample size and number of included studies, which make the findings stable and reliable and enable us to conduct multiple subgroup analyses by geographic location, study quality score, study design, etc. In addition, we also performed stratified analyses on the association of breastfeeding and risk of the subtypes of individual cancers.However, the present study has several limitations. First, there was high evidence of heterogeneity across studies in the categorical meta-analysis. However, meta-regression analyses suggested that geographic region and definition of reference category are the potential sources of the observed heterogeneity. Second, our meta-analysis included very limited studies from Oceania and no study from Africa. Third, the results of dose-response meta-analysis were prone to be influenced by possible exposure misclassification as the exposure dose was estimated with median for interval exposure, and the lower bound added to the half of the adjacent previous category for the highest open-ended exposure group. Fourth, the number of studies evaluating the associations of breastfeeding with risk of neuroblastoma, nephroblastoma, retinoblastoma, and germ cell tumors is small. Fifth, we were unable to assess differences by age or sex, because sufficiently age- or sex-specific studies are not available. Finally, limited by the lack information of breastfeeding pattern in most of included studies, such as exclusive breastfeeding and partial breastfeeding, we cannot evaluate the association between breastfeeding pattern and the risk of childhood cancer.This meta-analysis demonstrates that breastfeeding was associated with the reduced the risk of childhood leukemia. The present study also provides suggestive evidence of the inverse association between breastfeeding and risk of neuroblastoma. In addition, given that the role of breastfeeding for the risk of childhood leukemia and lymphoma may be region-specific, further analyses are warranted to provide insights into the strategy of breastfeeding advocacy.Additional file 1: Table S1. The PRISMA 2009 checklist for this meta-analysis. Figure S1. Flowchart of study selection. Table S2. List of excluded studies along with reason. Table S3. Details of included studies for subgroup analysis. Figure S2. Begg’s funnel plots identifying the publication bias for the association between breastfeeding and risk of childhood leukemia. Figure S3. Forest plots of subgroup analysis of association between breastfeeding and childhood leukemia risk in the order listed in Table Table S4. Subgroup analyses of the association between breastfeeding and acute lymphoblastic leukemia risk. Figure S4. One-study-removed analysis on the association of breastfeeding with risk of (A) childhood leukemia and (B) acute lymphoblastic leukemia. Figure S5. Pooled analysis of studies including only children aged 0-14 years old for the for the association of breastfeeding with risk of (A) childhood leukemia and (B) acute lymphoblastic leukemia. Figure S6. Risk estimates (solid line) and the corresponding 95% CIs (dash lines) for the dose-response relationship between breastfeeding and the risk of childhood lymphoma. Figure S7. Begg’s funnel plots identifying the publication bias for the association between breastfeeding and risk of childhood lymphoma. Figure S8. Forest plots of subgroup analysis of association between breastfeeding and childhood lymphoma risk in the order listed in Table Figure S9. One-study-removed analysis on the association of breastfeeding with risk of childhood lymphoma. Figure S10. Pooled analysis of studies including only children aged 0-14 years old for the for the association of breastfeeding with risk of childhood lymphoma. Figure S11. Begg’s funnel plots identifying the publication bias for the association between breastfeeding and risk of childhood brain tumors. Figure S12. Forest plots of subgroup analysis of association between breastfeeding and risk of childhood brain tumors in the order listed in Table Figure S13. One-study-removed analysis on the association of breastfeeding with risk of childhood brain tumors. Figure S14. Pooled analysis of studies including only children aged 0–14 years old for the for the association of breastfeeding with risk of childhood brain tumors.


The collagen 3 alpha 1 (COL3A1) rs1800255 polymorphism has been reported to be associated with women pelvic organ prolapse (POP) susceptibility, but the results of these previous studies have been contradictory. The objective of current study is to explore whether COL3A1 rs1800255 polymorphism confers risk to POP.Relevant literatures were searched by searching databases including Pubmed, Embase, Google academic, the Cochrane library, China National Knowledge Infrastructure (CNKI). Search time is from database foundation to March 2021.A total of seven literatures were enrolled in the present meta-analysis, including 1642 participants. Overall, no significant association was found by any genetic models. In subgroup analysis based on ethnicity, significant associations were demonstrated in Caucasians by allele contrast , homozygote comparison , and recessive genetic model .The present meta-analysis suggests that the COL3A1 is a candidate gene for POP susceptibility. Caucasian individuals with A allele and AA genotype have a higher risk of POP. The COL3A1 rs1800255 polymorphism may be risk factor for POP in Caucasian population. Pelvic organ prolapse (POP) is a phenomenon that pelvic organs such as uterus, vagina, bowel and bladder leave out of their normal anatomical location, resulting in a serious of functional disorders . It is aMany people attributed it to some risk factors including aging, obesity, vaginal delivery, hypoestrogenism, preoperative hysterectomy, nutrition, an many other environmental factors –9; howevAs is well known, the occurrence of POP is closely related to the integrity of pelvic floor support structure. The supportive structure of the pelvic floor depends largely on the extracellular matrix collagen. Of all the collagen, type I and type III are foremost structural fibers, which are closely associated with sidewalls of uterus, vagina and pelvic. The increased level of type III collagen is found in women who were suffering from POP . During Type III collagen, encoded by COL3A1 gene, consists of three identical α1 (III) chains . The genhttps://www.ncbi.nlm.nih.gov/pubmed/), EMBASE (https://www.embase.com/), the Cochrane library(https://www.cochranelibrary.com/), Google academic (https://scholar.google.com/), and Chinese national knowledge internet (https://www.cnki.net/). Search time is from database foundation to March 2021. The keywords applied in the search process were as follows: “polymorphism” or “polymorphisms” or “genome wide association study” or “genome-wide association studies” or “GWAS” or “rs1800255”, “Pelvic organ prolapse” or “POP”. The literature language was limited to English language and Chinese language. Additionally, in order to avoid the omission of relevant literatures, we searched the references as much as possible.Two independent authors systematically and comprehensively searched Pubmed (The inclusion criteria must meet a series of conditions: (a) a case-control study; (b) making an assessment of association between COL3A1 rs1800255 polymorphism and POP risk; (c) offering sufficient data or information to count OR and 95%CI. The literatures are excluded when they fit one or more of these factors of exclusion criteria: (a) case reports, reviews, meta-analysis, letters, and editorials; (b) not offering sufficient data for the present meta-analysis. (c) its experiment objective was pig, rat or other animals.All the necessary information was independently searched, reviewed and assessed by two authors . Then this contradictory data or information was reassessed by the corresponding author (Yongxian Lu). The extracted data consisted of author name, publication year, genotyping methods, sample size, ethnicity, matching criteria, source of control, HWE conformity; Newcastle-Ottawa Scale (NOS) score . If the 2 statistics would be applied [2 <50% was considered to low heterogeneity, 50 ≤I2 <75% was considered to moderate heterogeneity and I2 ≥75was considered to significant heterogeneity. Furthermore, the Galbraith plot was used to spot the outliers to find out the potential heterogeneity as much as possible. Sensitive analysis was applied to detect the influential studies which might contribute obvious bias to final results. The funnel plot and Egger’s test were put into use to recognize the existence of publication bias [The association power was assessed through the corresponding indexes including OR and 95%CI. And both the Q-statistic and I applied . Four ge applied , 26. I2 ion bias . The Staion bias .The flow chart of meta-analysis search course was shown in PRISMA 2009 Flow Diagram . Based oThe meta-analysis results between COL3A1 rs1800255 polymorphism and POP susceptibility are shown in 2 = 14.89, P = 0.021, I2 = 59.7, 2 = 16.4, P = 0.012, I2 = 63.4, 2 = 16.04, P = 0.171, I2 = 62.6, 2 = 9.04, P = 0.014, I2 = 33.6, 2 = 0, 0, 0, 0). However, there was still significant heterogeneity in Asian population under all genetic models . In order to further detect the source of heterogeneity, we conduct Galbraith plots to find out the outliers which might influence the heterogeneity. Consequently, we found that all the studies were within reasonable limits except Jeon et al  chains. The genetic polymorphisms of COL3A1 gene lead to amino acid change of the α1 (III) chain. This COL3A1 rs1800255 polymorphism is a G/A variation and located in position 2209 (exon 30). The above exchange leads to change of protein-level expression . And threonine has a larger and more hydrophilic side chain than alanine, leading to disturb the triple helical conformation of type III collagen. In recent years, N4-acetylcytidine (ac4C) has become a research hotspot and has been paid wide attention. Recent studies have shown extensive ac4C modifications in human and yeast mRNAs . And ac4This is, to our knowledge, the first meta-analysis of genetic studies on the association between POP risk and COL3A1 rs1800255 polymorphism. We found that COL3A1 rs1800255 polymorphism contributes no risk to POP of Brazilian and Asian population but contributes increased risk to POP of Caucasian population. To be specific, the A allele and AA genotype were risk factors and they contribute a increased risk to POP risk in Caucasian population. Eliminating the source of bias is of vital importance for gene polymorphism association meta-analysis. Hence, we have attempted to conduct all the three patterns in the present meta-analysis. Firstly, allele contrast was used to find out the high risk or low risk allele. Secondly, homozygote comparison was used to find out the high risk or low risk genotype. The last pattern is comparing homozygote genotype versus allele carriers. In the present study, the low-moderate heterogeneity between studies occurred in the overall population. Common reasons for heterogeneity consist of differences in the investigated populations or in genotyping methods or in sample size or it may be derived from other risk factors. By performing meta-regression, subgroup analysis and Galbraith plot analysis, we found that ethnicity and sample size might contribute substantial heterogeneity to final results. In the present study, we successfully find out the source of heterogeneity. And the results of sensitive analysis and publication bias demonstrated that the results of our meta-analysis were stable and reliable.Some limitations should be pointed out. Firstly, variable confounding factors including aging, obesity, vaginal delivery, hypoestrogenism and preoperative hysterectomy are not fully considered due to limited dataset , 42, theIn summary, the present meta-analysis suggests that the COL3A1 is a potential risk gene for POP. Caucasian individuals with COL3A1 rs1800255 A allele and AA genotype have a higher risk of POP. The COL3A1 rs1800255 polymorphism may be a risk factor for POP in Caucasian population. Further studies investigating other confirmed genetic factors and possible gene-gene and gene-environmental interactions for COL3A1 rs1800255 polymorphism in a large GWAS data should be warranted.S1 Table(DOC)Click here for additional data file.S1 File(DOC)Click here for additional data file.


Background: An association between lower urinary tract symptoms (LUTS) and risk of sexual dysfunction in male remains controversial in recent decades.Materials and Methods: PubMed and Web of Science were searched up to October 28, 2020, for articles reporting the prevalence of sexual dysfunction in men with LUTS. The main outcomes were results from sexual dysfunction assessments. Pooled odds ratio (OR) and weighted mean difference (WMD) with 95% confidence interval (CI) were calculated. The quality assessment of the included studies was performed by using The Newcastle-Ottawa Scale (NOS) or JBI Meta-Analysis of Statistics Assessment and Review Instrument (JBI-MAStARI).Results: A total of 24 full-manuscript papers met the inclusion criteria. The pooled OR for 21 studies suggested that patients with severer LUTS had a higher risk of sexual dysfunction . A significant decrease in scores of assessment tools for sexual dysfunction was observed in the patients with higher severity of LUTS compared with those patients with lower severity . Similar outcomes were also found in subgroup analyses. In a detailed analysis of specific sexual function domains, the severity of LUTS was associated with erectile dysfunction, intercourse satisfaction, and overall satisfaction, except for sexual desire.Conclusion: The study demonstrates an association between exposure of lower urinary tract symptoms and risk of sexual dysfunction in male. Assessment of sexual function is necessary for patients with lower urinary tract symptoms.Systematic Review Registration:http://www.crd.york.ac.uk/prospero, identifier: CRD42020208747. Lower urinary tract symptoms (LUTS) are common in elderly men and the prevalence increases with age according to the epidemiological studies . LUTS inSexual dysfunction is defined as difficulty experienced by an individual or a couple during any stage of a normal sexual activity which results in misery and strained interpersonal relationship. Sexual dysfunction includes sexual desire disorders, sexual arousal disorders, orgasmic disorders, and sexual pain disorders. Erectile dysfunction belongs to sexual arousal disorder , 8. SexuSome epidemiological studies have shown that there is an association between exposure of LUTS and risk of sexual dysfunction , 12. Undwww.crd.york.ac.uk/prospero).This work was executed in accordance with the Preferred Reporting Items for Systemic Reviews and Meta-analysis (PRISMA) guidelines. According to PRISMA guideline, in current study, the populations were adult male, the exposures were individuals with LUTS or higher severity of LUTS, the comparators were individuals without LUTS or with lower severity of LUTS, the outcomes were results of sexual dysfunction assessment. We have registered this meta-analysis in PROSPERO  Studies reported the association between LUTS and sexual dysfunction. LUTS refer to a group of clinical symptoms involving the bladder, urinary sphincter, urethra and the prostate in men, which included increased frequency of urination, increased urgency of urination, urge incontinence, excessive passage of urine at night, poor stream, hesitancy, terminal dribbling, incomplete voiding, urinary retention, overflow incontinence and episodes of near retention. Sexual dysfunction included sexual desire disorders, sexual arousal disorders, erectile dysfunction, premature ejaculation, orgasm disorders, sexual pain disorders and post-orgasmic diseases; (2) Studies reported the prevalence or the number of sexual dysfunction patients, or the average scores with standard deviation of assessment tools for sexual function; (3) Clinical studies were performed with adult males and published in English. The exclusion criteria were as follows: (1) The study type was a review, case report, abstract of conference, comment or editorial. (2) Study without appropriate comparator. (3) Study reported incomplete data of outcomes, such as the lack of standard deviation of sexual function assessment score. (4) The determination of LUTS and ED didn't base on appropriate clinic diagnosis or relevant questionnaires. The questionnaires for LUTS assessment included International Prostate Symptom Score (IPSS) questionnaire, Danish Prostatic Symptom Score (DAN-PSS-1) questionnaire and The National Institutes of Health Chronic Prostatitis Symptom Index (NIH-CPSI). The questionnaires for sexual function assessment included International Index of Erectile Function (IIEF), the Brief Male Sexual Function Inventory (BSFI), Danish Prostatic Symptom Score (DAN-PSS-sex) and Epstein Inventory. The details of these questionnaires are provided in the The titles and abstracts were screened independently by two authors, and then the full texts of the relevant studies were reviewed later. The reference lists of relevant articles were hand-searched. Two authors independently extracted and cross-checked the following data: first author, publication year, nation, age, study type, sample size, criteria of LUTS severity categories, assessment tools for LUTS and sexual function, and diagnosis criteria of sexual dysfunction. Any disagreements were discussed by the two authors or sought help from the third author.The quality of case-control and cohort studies was assessed using The Newcastle-Ottawa Scale (NOS), which has a maximum score of 9. Studies with a total score of 1 to 3, 4 to 6, 7 to 9 in the NOS scale were considered low, intermedia, and high quality, respectively. The quality of cross-sectional and longitudinal studies was assessed using JBI Meta-Analysis of Statistics Assessment and Review Instrument (JBI-MAStARI) . The JBII2 statistic. A P-value > 0.1 or an I2 statistic ≤ 50% suggested low heterogeneity across studies, and the fixed effect model was chosen. Otherwise, the random effect model was applied. For dichotomous data, the odds ratio (OR) with 95% confidence interval (CI) was used, while the weighted mean difference (WMD) with 95% CI was used for continuous data. P-value <0.05 were considered statistically significant.The meta-analysis was conducted using Review Manager 5.4 software. The heterogeneity across studies was tested by the Q statistic and The flow diagram of the searching and screening process was outlined in The sample sizes of the included studies ranged from 63 to 2,916. Detailed characteristics were listed in p < 0.001, I2 = 90%, P < 0.001).Twenty-one studies –35 reporP < 0.001, I2 = 96%, P < 0.001).Six studies , 36–38 rP = 0.002, Four studies reported the erectile function domain in their results , 36, 38.P = 0.07, Four studies evaluated the sexual desire of patients with different severity of LUTS , 36, 38.P = 0.001, Four studies evaluated the intercourse satisfaction domain , 37, 38.P < 0.001, Three studies evaluated the overall satisfaction domain , 27, 36.Subgroup-analysis was performed according to age, assessment tools for LUTS and assessment tools for sexual function.P = 0.002) and participants with age >40 years . Inter-study heterogeneity was high and significant in both groups . The pooled analysis forest plot is shown in In subgroup-analysis, the results indicated that the prevalence of sexual dysfunction in participants with higher severity of LUTS was higher compared with participants with lower severity of LUTS in both participants with age ≤40 years . Analysis of the study that used clinical diagnosis to assess LUTS revealed that there was no association between exposure of LUTS and risk of sexual dysfunction . The pooled analysis forest plot is shown in To assess the severity of LUTS, IPSS was used in 15 studies , 32, 35,P < 0.001, I2 = 92%; IIEF group: WMD = −5.80, 95%CI: −9.70 to −1.91, P = 0.003, I2 = 37%; BMSFI group: WMD = −2.60, 95%CI: −3.10 to −2.10, P < 0.001; Epstein inventory group: WMD = −5.98, 95%CI: −6.75 to −5.21, P < 0.001). The pooled analysis forest plot is shown in In those studies which reported scores of assessment tools for sexual dysfunction, 4 assessment tools had been used to measure sexual dysfunction. Five-item International Index of Erectile Function (IIEF-5), IIEF, Brief Male Sexual Function Inventory (BMSFI) and Epstein Inventory were used to assess sexual dysfunction in two studies , 37, twoIn our meta-analysis, the result revealed that participants with higher severity of LUTS had a higher prevalence of sexual dysfunction compared with those participants with lower severity of LUTS. Specifically, the result indicated that participants with higher severity of LUTS had worse erectile function, intercourse satisfaction, and overall satisfaction compared with those participants with lower severity of LUTS. Therefore, LUTS may impact sexual activity including erectile function and sexual satisfaction.According to previous studies, age may be a significant confounding factor of sexual dysfunction . However1-adrenoceptors and the secretion of noradrenaline, which can mediate adrenergic contraction of smooth muscles in the bladder neck, prostate, urethra, and the corpus cavernosum  is expressed in the whole of the lower urinary tract, including the urethra, prostate, and bladder . PDE5 isreatment .The most important strength of this meta-analysis is that the result may be a reliable evidence to indicate the association between LUTS and sexual dysfunction. This meta-analysis is the first to investigate the association between LUTS and sexual dysfunction. Therefore, the result is useful for clinicians, policymakers and patients. For patients with sexual dysfunction, the probability of concurrent LUTS should be considered and assessed. The treatment of LUTS may help the treatment of sexual dysfunction.There are also some limitations. First, we only searched the literature in two databases and some relevant papers in other databases, such as Embase, may be missed. Although we searched Embase after original systematic review and found no additional suitable article. Second, although subgroup analyses were conducted, the high heterogeneity due to the differences of assessment tools among studies made it difficult to generalize the conclusion. Furthermore, the study types of the included studies only include cross-sectional study and longitudinal study, so the result of the meta-analysis cannot determine the causality between LUTS and sexual dysfunction. Further cohort and case-control studies should be performed to investigate the relationship between LUTS and sexual dysfunction.Males with higher severity of LUTS had worse sexual function compared with those with lower severity of LUTS. The result also gives a remind to clinicians, policymakers, and patients that exposure to LUTS may also have a high probability of sexual dysfunction, and they should pay attention to the accidence of sexual dysfunction.The original contributions presented in the study are included in the article/GS and MW contributed to data acquisition. XL and YZ came up with the idea of doing this research and are responsible for the whole work. GS, MW, BC, GL, HL, RL, ZL, CW, TW, SW, and JL participated in data analysis, writing, and revision of the article. All authors read and approved the final version of the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest. The reviewer XY declared a shared affiliation, with no collaboration, with the authors to the handling editor at the time of the review.


Research evidence has established the beneficial effects of diet in cancer prevention; various epidemiological studies have suggested that olive oil component could play a role in decreasing cancer risk. This systematic review and meta-analysis aims to investigate the association between olive oil consumption, cancer risk and prognosis.A systematic search was conducted in PubMed, EMBASE and Google Scholar databases . Pooled relative risk (RR) and 95% confidence intervals (95% CIs) were estimated with random-effects (DerSimonian-Laird) models. Subgroup analyses, sensitivity analyses and meta-regression analysis were also performed.45 studies were included in the meta-analysis; 37 were case-control  and 8 were cohort studies . Highest olive oil consumption was associated with 31% lower likelihood of any cancer , breast , gastrointestinal , upper aerodigestive  and urinary tract cancer . Significant overall effects spanned both Mediterranean and non-Mediterranean participants, studies presenting a multivariate and a univariate analysis and all subgroups by study quality.Olive oil consumption seems to exert beneficial actions in terms of cancer prevention. Additional prospective cohort studies on various cancer types and survivors, as well as large randomized trials, seem desirable. Cancer is accountable for an estimated 9.6 million deaths in 2018, being the second leading cause of death globally, only after cardiovascular diseases . AccordiOlea europaea, Oleaceae) is a traditional staple food for Mediterranean people and a fundamental component of the Mediterranean diet, used for both dressing and cooking. It has the highest ratio of monounsaturated to polyunsaturated fatty acids among vegetable oils. Its favorable effects have been attributed to the abundance of valuable nutrients, such as antioxidant phenolic compounds , vitamins, lignans, squalene and terpenoids , definition of olive oil intake, categorization of exposure, features of ascertainment for exposure and characteristics of participants , as well as adjusting factors regarding multivariate analyses. If the required data for the meta-analysis were not readily available in the published article, the corresponding authors were contacted twice (a reminder e-mail was sent seven days after the first e-mail). Data were independently extracted, analyzed and recorded in a predeveloped data extraction sheet by two reviewers (CM and MO). Final decision was reached after consultation with a senior author (TNS) and team consensus.The maximally adjusted effect estimates i.e., odds ratios (ORs) for case–control studies, relative risks (RRs) or hazard ratios (HRs) in case of randomized controlled trials and non-randomized cohort studies with their confidence intervals (CIs) were extracted from each study by category of olive oil intake. In case the aforementioned information was not available, crude effect estimates and 95% CIs were calculated by means of 2x2 tables presented in the articles.post hoc meta-regression and sensitivity analyses. Statistical synthesis was performed in case of two or more eligible study arms. Random-effects (DerSimonian–Laird) models were appropriately used to calculate pooled effect estimates. The category of highest olive oil intake was compared with the one corresponding to the lowest consumption. Between-study heterogeneity was assessed by estimating Q-test and I2 [post hoc subgroup analysis was performed in studies examining olive oil consumption within or outside the context of adherence to Mediterranean diet. Post hoc sensitivity analyses were performed excluding the following effect estimates; first those pertaining to cancer mortality (and not purely incidence) and second, in the site-specific analyses, a study incorporating in the pool of upper aerodigestive cancer also esophageal cancer cases [Statistical analyses included pooling of studies as well as t and I2 . Separater cases . Meta-reer cases  and aimea priori at 85% response rate and 5 years, respectively. Study quality was considered “low” when the Newcastle-Ottawa score (NOS) ranged between 1–3, “intermediate” for studies with NOS between 4–6 and “high” for those with a score between 7–9. Two independently working reviewers  rated the studies and, in case of disagreement, final decision was reached after consultation with a senior author (TNS) and team consensus.As far as the risk of bias is concerned, the Newcastle-Ottawa Quality scale  was usedPublication bias was evaluated in the analyses that included 10 or more study arms ; Egger’sA total of 3813 records were identified  using the search algorithm. After duplicates were removed, 2413 abstracts were screened; all details pertaining to the successive steps for the selection of eligible studies are provided in the supplemental material  [Two studies investigated only the effect of incremental olive oil consumption , 71 and,atients) . The eva45 eligible studies were included in the overall meta-analysis; 37 were case-control  and 8 were cohort studies  , Table 1post hoc subgroup analysis showed similar results in studies examining olive oil consumption in the context of adherence to Mediterranean diet  and outside that context . The results persisted in the post hoc sensitivity analysis excluding studies on cancer mortality  ; a non-significant trend was observed in cohort studies  , Fig 2. As far as breast cancer is concerned, pooling of 14 study arms resulted in a protective association . Similarly to the analysis on overall cancer, the beneficial effect was reproducible in case-control  but not in cohort studies , Fig 3. The risk for gastrointestinal cancer was found to be 23% lower for those who consumed the highest amounts of olive oil . When we proceeded with subanalyses per tumor site, an inverse relationship was found between olive oil intake and risk for esophageal (RR = 0.47 95%CI: 0.24–0.93) and pancreatic cancer  , Fig 4; In the case of upper aerodigestive tract cancers, favorable effects of higher olive oil consumption were found  ; subgrouRemarkably, pooled analysis on urinary tract cancers indicated a relative risk of 0.46 (95%CI: 0.29–0.72) ; examinaThe relevant passage is presented in the S1 File and S10 Table in The evaluation of quality within the eligible studies is presented in S6 and S6 Tables in In case-control studies, hospital-based controls often compromised quality. Cases were representative in all 38 studies and were defined adequately in all but one studies. In terms of exposure, many studies contained no description of non-response rates. Nevertheless, the same method of ascertainment was uniformly guaranteed for both cases and controls in all studies; standardized, validated food frequency questionnaires were used through a structured interview for most studies.via Egger’s test in the analysis on overall cancer risk (p<0.001), breast cancer (p = 0.013) and gastrointestinal cancer risk (p = 0.048). These results are reflected as asymmetry in the respective funnel plots  for the highest versus lowest level of olive oil consumption after evaluating five case-control and one cohort study [et al. reported a pooled effect estimate of nine case-control and three cohort studies of 0.74 (95%CI: 0.60–0.92) [et al. included also articles on the use of monounsaturated fatty acids as well as olive oil combined with frying/liquid oils.Increasing evidence supports that olive oil constituents convey protection against the development of several types of cancer , 83. Thert study . Focusin60–0.92) , with thin vivo and human in vitro studies. The favorable effect of olive oil is largely attributed to its exceptional composition, rich in monounsaturated fatty acids (mainly oleic acid) [via scavenging and influence crucial signaling pathways linked to carcinogenesis [in vitro studies indicated that oleic acid is able to transcriptionally repress Her-2/neu overexpression and to upregulate PEA3, a transcriptional repressor of the HER2 gene [From a mechanistic point of view, our findings comply with several experimental ic acid) , squalenic acid) , 85. Theogenesis . RegardiER2 gene . It has ER2 gene . In humaER2 gene  and squaER2 gene  were fouER2 gene .et al. reported that both free and lipid bound phenolic extracts of virgin olive oil exhibited antiproliferative activities against the colorectal cancer cell lines CRC1 and CRC5 [et al. the extracts limited invasion in vitro and metastasis in vivo more likely via modulation of integrin expression [Regarding gastrointestinal effects of olive oil, a plethora of preclinical evidence points to a protective role of its components . In humaand CRC5 , whereaspression . Additiopression .Commenting on effects in the urinary tract, EVOO phenolic extract suppressed proliferation and clonogenic ability in a dose-dependent manner in human urinary bladder cancer cells (T24 and 5637) . Oleuropet al., strongest adherence to a Mediterranean diet was inversely associated with cancer mortality and risk of various cancer types; nevertheless, pooled data about the use of olive oil as a single component pointed to a non-significant effect on overall cancer risk, synthesizing a subset of relevant studies [According to the recent meta-analysis by Schwingshackl  studies . High olRegarding the limitations of the present meta-analysis, between studies heterogeneity in the overall analysis was substantial but in line with previous meta-analyses . HeterogNext, 30 out of 45 studies were limited to Mediterranean populations, where olive oil is the core of the diet, whereas none was detected in the American area, thus, compromising the generalizability of the results. Moreover, details about different types of olive oil examined  were not provided by the individual studies, as a rule; given that concentrations of polyphenols, antioxidants and anti-inflammatory compounds may well differ across different types of olive oil, the fact that the reporting in synthesized studies did not allow the performance of relevant subgroup analyses represents another limitation of this meta-analysis. Finally, some cancer types were not studied while others  did not generate enough eligible study arms to allow for further subgroup analyses; regarding breast cancer there was paucity of data about differential effects by menopausal status and expression of hormone receptors. Only one study  examinedper se and not as a source of monounsaturated fatty acids or mixed with other components. Furthermore, available information was depicted on a considerable set of meaningful subanalyses and sensitivity analyses, where the favorable effects of olive oil were frequently persisted.Despite the above mentioned limitations, the present work possesses a plethora of important strengths. First of all, our updated search was performed in three online databases that cover the most of biomedical literature and it was not subject to any restriction. Moreover, through strict and meticulous adherence to the PRISMA guidelines  as well In conclusion, the results of this meta-analysis represent valuable evidence of the protective effects of olive oil against cancer development. Additional prospective cohort studies on various cancer types, especially in non-Mediterranean regions, as well as large randomized trials, seem desirable in order to provide further insight into the role of olive oil in preventing cancer.S1 File(RAR)Click here for additional data file.


Although several epidemiological studies have attempted to evaluate the relationship between cholecystectomy and gastric cancer risk, the findings have been controversial. This study aimed to carry out a systematic review and meta-analysis following the reporting guidelines to comprehensively analyze and quantify the evidence of the aforementioned association.Studies were identified by searching the Medline (PubMed), Embase, and Web of Science from inception to November 30, 2020, with only studies published in English being considered. Summary relative risks (RRs) and 95% confidence intervals (CIs) were calculated by random-effects models.P = 0.117, I2 = 37.8%). These positive findings were consistent in most subgroup analyses like region in Asia, number of cases ≥200, cohort study design, sex in male, low risk of bias, exposure collection by database, and adjustments made for age, gender, calendar year. Of note, we also observed positive association between cholecystectomy and non-cardia of gastric cancer risk . No publication bias was present.Eight studies (five cohort studies and three case–control studies) with a total of 26,063 gastric cancer patients and 848,081 participants were included. The summarized RR of the relationship between cholecystectomy and gastric cancer risk was 1.11 (95%CI: 1.03–1.20), with low heterogeneity (The aforementioned evidence suggested that a history of cholecystectomy was associated with a slightly elevated risk of gastric cancer. Results of most subgroup analyses also supported the main findings. More prospective studies are warranted to further validate these findings. Helicobacter pylori infection, consumption of foods preserved by salting, low intake of fruits, alcohol consumption, and active tobacco smoking (Gastric cancer (GC) is one of the most common cancers in the world. Although the incidence of GC is decreasing, it is still the sixth most common malignancy and the third leading cause of cancer-related deaths. The incidence of GC is significantly elevated in East Asia, while that in North America, northern Europe, and the entire African region is generally lower , 2. GC i smoking –5.Cholelithiasis  is the most common gastrointestinal disease. An estimated 10% of Europeans and Americans and 5–10% of Asians are carriers of gallbladder stones , 7. The Considering that the previous meta-analysis is out of date and does not include research from the past decade, and given the conflicting conclusions of current research, this study aimed to carry out an updated systematic review and meta-analysis of observational studies to evaluate the association between cholecystectomy and the risk of GC.The reporting standards of the Meta-Analysis of Observational Studies in Epidemiology group  and PrefTwo independent individuals (YY and M-HL) comprehensively searched the Medline (PubMed), Embase, and Web of Science from inception to November 30, 2020, with only studies published in English being considered. Details of the full search strategy are provided in the A study was eligible for inclusion if it (1) utilized an observational study design; (2) evaluated the relationship between cholecystectomy and GC risk; and (3) demonstrated estimates of odds ratios (ORs), relative risks (RRs), or hazard ratios (HRs) with 95% confidence intervals (CIs) or extractable data necessary to calculate these parameters. However, publications meeting any of the following criteria were excluded: (1) clinical trials, letters, editorials, case reports, reviews, meta-analyses, and meeting abstracts; (2) lack of sufficient risk estimates or related data to calculate risk estimates; and (3) not published in English. The selection and exclusion of studies were reviewed by two investigators (YY and M-HL). Disagreements were resolved by consensus with a third author (YL).Data were extracted in duplicate using standardized forms. Disagreements were resolved by consensus. The following information was collected: last name of the first author, publication year, geographical location, study design, number of cases, number of controls/cohorts, and characteristics of exposure and covariates matched in the study design or adjusted in the statistical analysis.Quality assessment was performed using the Newcastle–Ottawa Scale (NOS), which consisted of eight items grouped into three domains  to assess the methodological quality of case–control or cohort studies . StudiesI2 statistics. 2;I estimates the proportion of variability in the meta-analysis caused by differences between studies instead of sampling error . Fewer studies were adjusted for the calendar year (n = 3). Specifically, one study was adjusted for more than five potential confounders in the primary analysis.The characteristics of the eight included studies are summarized in I2 = 37.8%). No publication bias was present  (The estimates by subgroups together with the results of the heterogeneity tests are given in I2 = 43.5%) to 1.13  , the probability of chance findings could not be ruled out. Additionally, although the two included studies both supported the aforementioned positive correlation results, Fall et al.  conducteThe underlying exact mechanisms of these contradictory links between cholecystectomy and GC risk have been unclear; however, some potential plausible mechanisms have been proposed to explain these findings. After cholecystectomy, bile flow changes, increasing the bile exposure of the stomach, changing bile salts, and subsequently changing the levels of metabolic hormone . IncreasTo our knowledge, the present study is the most comprehensive meta-analysis of cholecystectomy and GC risk so far. The strengths of this study include the following: First, this meta-analysis involved a large sample size  to evaluate the effect of cholecystectomy and GC risk, which increased the statistical power to detect the association. Second, numerous subgroup analyses were performed to analyze the study characteristics that might affect results, and sensitivity analyses were further performed to explore the heterogeneity in this study. Third, this present meta-analysis had no publication bias and low heterogeneity, and most of the included studies had a low risk of bias. All these strengths make the results of this study more convincing.H. pylori infection (However, the study also had several limitations. First, a significant excess risk of GC related to cholecystectomy was observed in cohort studies, but not in case–control studies. Furthermore, the subgroup analysis of case–control studies represented moderate heterogeneity, which was higher than the subgroup of cohort studies. The case–control studies were prone to generate selection and recall bias, and the quality of the cohort studies was inconsistent, which might explain the observed heterogeneity in the study. Second, the different exposure rates of cholecystectomy varied among the included studies, which might be an important issue. Most included studies were conducted in European countries, and only a few were conducted in other countries. The exposure rate of cholecystectomy was observed to be 20% of the total participants in Taiwan but 10% of that in the United Kingdom , 31. Thinfection  and dietnfection . Fourth,nfection . The patnfection . TherefoThe present systematic review and meta-analysis indicated that cholecystectomy had an increased risk of developing GC. Meanwhile, most subgroup analyses also supported the main findings. More large-scale prospective cohort studies are needed to validate these findings worldwide to gain further insights.yanli0227@126.com.The data that support the findings of this study are available from the corresponding author upon reasonable request. Requests to access the datasets should be directed to YL: conceived and designed the study. YY and M-HL: literature search, data curation and formal analysis, writing the original draft. All authors contributed to the article and approved the submitted version.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. 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Background: Total knee arthroplasty (TKA) is a commonly performed procedure, primarily when knee joints have been damaged by progressive arthritis; however, over 20% of surgical patients develop persistent postsurgical pain (PPSP). We plan to conduct a systematic review and meta-analysis of factors associated with the development of PPSP following TKA.Methods: We will include peer-reviewed cohort or case–control studies that explore, in an adjusted model, factors associated with the development of PPSP after TKA. We will identify eligible studies, in any language, by a systematic search of MEDLINE, EMBASE, CINAHL, AMED, Scopus, SPORTDiscus, and PsycINFO, from inception of each database. Pairs of reviewers will, independently and in duplicate, screen titles and abstracts of identified citations, review the full texts of potentially eligible studies, and extract information from eligible studies. When possible, we will pool estimates of association for all independent variables reported by more than one study and report both an adjusted odds ratio and the absolute risk increase and associated 95% confidence intervals (Cis). We will use the Grading of Recommendations Assessment, Development and Evaluation (GRADE) approach to summarize the quality of evidence for all meta-analyses as high, moderate, low, or very low.Discussion: Our results will facilitate identification of patients at risk for the development of PPSP following TKA, highlight promising predictors for further study, and help guide the design of interventional studies to improve prognosis of high-risk patients. The other review that provided pooled estimates only looked at procedural and surgical techniques as predictors of anterior knee pain following primary TKA, combined both unadjusted and adjusted estimates of association, and pooled treatment effects from randomized controlled trials with measures of association from observational studies.19 We propose to conduct a new systematic review and meta-analysis of observational studies to identify predictors of PPSP following TKA that addresses limitations of prior reviews.We found eight reviews that have explored predictors of PPSP following TKA, of which three were narrativeeviews19 . The sys20We registered our protocol with PROSPERO (CRD42018065943) and will follow the Meta-analysis of Observational Studies in Epidemiology Statement for reporting of our systematic review.We will systematically search MEDLINE, EMBASE, CINAHL, AMED, Scopus, SPORTDiscus, and PsycINFO, from inception of each database, without any language restriction. An experienced medical librarian (R.C.) has developed search strategies for each database (Appendix). We will review reference lists from eligible studies and related reviews for additional potentially eligible studies.We will include peer-reviewed cohort and case–control studies that enroll adults (18 years or older) who undergo TKA and investigate, in an adjusted analysis, risk factors for PPSP after TKA. Studies will be ineligible if their predictive models include significant associations with variables collected after baseline, because the association may be a result of PPSP. When study populations overlap by more than 50% among eligible articles, we will include only the study with the largest sample size and longest follow-up. We will exclude conference abstracts.https://www.evidencepartners.com/products/distillersr-systematic-review-software/) to facilitate literature screening and prepare aPreferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) flow diagram to illustrate the flow of studies through the selection process.Pairs of reviewers will screen the titles, abstract, and full-text articles of potentially eligible studies, independently and in duplicate. Reviewers will, when necessary, resolve disagreements by discussion or by consultation with an adjudicator. We will use online systematic review software , Canada; Pairs of reviewers, working independently and in duplicate, will extract relevant information from all eligible studies. Before starting data abstraction, we will conduct calibration exercises to ensure consistency between reviewers. We will use piloted, standardized forms to extract the following information: (1) study characteristics ; (2) study population characteristics ; (3) surgical procedure ; (4) risk of bias and statistical analysis approaches; and (5) measures of association with PPSP for all independent risk factors explored in an adjusted model.When a study reports more than one regression model, we will use the model with the largest population, longest follow-up, or largest number of risk factors. When studies report the dependent variable (PPSP) as both a dichotomous outcome  using logistic regression and as a continuous outcome  using linear regression, we will use the results from the logistic regression model. All disagreements on data extraction will be resolved through discussion. We will contact authors for clarification of eligibility or missing data.Users’ Guides to the Medical Literature to assess the risk of bias: (1) representativeness of study population; (2) validity of outcome assessment; (3) proportion of missing data and loss to follow-up (≥20% will be considered high risk of bias); and (4) whether predictive models are appropriately adjusted.21 We define a model as appropriately adjusted when it includes age, sex, and a measure of disease severity . We will assess whether the adjusted model was data driven  or theory driven .We will use the following criteria from the 22 We will report intensity of PPSP across studies as the median and interquartile range, after converting all pain scales to a 10-cm visual analogue scale.24 For continuous predictors that are entered as categorical variables in the regression model , we will assume linearity and that the associations across categories are independent of each other and calculate the OR and 95% confidence interval (CI) for each category using Bucher’s approach and combine ORs using the inverse variance method to produce a single OR for the predictor.26 In studies that excluded predictors from their final adjusted analysis due to nonsignificant association in univariate or bivariate analysis or the adjusted OR and its 95% CI were not reported due to nonsignificant association in final adjusted model, we will use an OR of 1 and impute the associated variance using the hot deck approach to avoid overestimation of association.27We will assess interrater agreement of full-text screening with the kappa statistic.28 We will use DerSimonian-Laird random effects models for all meta-analyses.29 We will assess publication bias by visual assessment of funnel plots and Egger’s test when at least ten studies are included in a meta-analysis.31We will pool all independent factors assessed for an association with PPSP that are reported by more than one study as an OR and associated 95% CI and calculate the absolute risk increase. We will estimate the baseline risk for PPSP after TKA using the lowest rate of PPSP from the study eligible for review with the largest sample size among studies at low risk of bias. When the measure of association is reported as relative risk, we will convert it to an OR using the reported baseline risk in the reference or unexposed group (participants without the risk factor).P ≤ 0.01, (2) a large magnitude of association (OR ≥ 2.0), and (3) a sample size of at least 500 patients. Data analysis will be performed using STATA software (Version 15.1).If pooling is not possible, we will explore the consistency of association between pooled results and studies reporting the same predictors that could not be pooled. We will define nonpoolable predictors as promising if they meet the following criteria: (1) a statistically significant association with PPSP of 32 We will explore four a priori hypotheses to explain variability between studies, assuming a larger association with PPSP with the following study characteristics: (1) longer duration of follow-up, (2) higher threshold for PPSP , (3) larger proportion of patients lost to follow-up, and (4) greater risk of bias on a criterion-by-criterion basis. In addition, we will conduct a subgroup analysis exploring studies that adjusted for preoperative pain vs. those adjusted with another measure of disease severity ), and if we find a significant subgroup effect we will prioritize pooled measures of association from studies that adjusted for pain severity. We will conduct subgroup analyses only if each subgroup contains three or more studies and explore for subgroup effects with a test of interaction.33 We will examine the effect of imputing data for nonsignificant predictors and converting categorical data for patient age to continuous data in a sensitivity analysis.Statistical tests of heterogeneity can be misleading when estimates of precision are very narrow due to large sample sizes; thus, we will evaluate heterogeneity for all pooled estimates through visual inspection of forest plots.34 Once we have established the baseline risk of PPSP after TKA, we will estimate the absolute increase in risk for both modifiable and nonmodifiable factors that would alter clinical decision making and rate down for imprecision if the 95% CI associated with the risk difference includes this threshold.We will use the Grading of Recommendations Assessment, Development and Evaluation (GRADE) approach to assess the certainty of evidence as high, moderate, low, or very low, based on risk of bias, inconsistency, indirectness, imprecision, and publication bias.Due to the large volume of TKAs performed each year and the high rate of patients who develop PPSP, there is an urgent need for a high-quality systematic review to identify factors associated with the development of persistent pain. Modifiable factors that show large associations have the potential to be directly targeted to reduce the rate of PPSP after TKA. Moreover, patients who present with important nonmodifiable factors may benefit from nonspecific interventions. Either scenario would help direct clinical trials to establish the effectiveness of promising strategies.Our proposed review has several strengths in relation to prior reviews. First, we will update the search. Second, we will pool similar predictors across studies and optimize interpretability by reporting measures of association in both relative and absolute measures. Third, we will incorporate nonsignificant predictors into our meta-analyses by imputing measures of association when they are not reported to avoid overestimating the strength of associations. Fourth, we will assess the overall certainty of evidence using the GRADE approach.A potential limitation is the quality and comprehensiveness of reporting in the primary literature. We may find that nonmodifiable factors, such as disease severity or surgical approaches, have been commonly explored for their association with PPSP after TKA, whereas modifiable factors, such as anxiety or coping abilities, have not. The findings of our review will help inform patients considering TKA about their prognosis and identify key areas for future research.Click here for additional data file.