The burden of pancreatic cancer varies greatly across countries, with the number of deaths, incident cases, and disability-adjusted life years more than doubling in recent years, and with high-income countries having the highest incidence and mortality rates. We conducted this systematic review with meta-analysis with the goal of summarizing the current evidence on dietary fiber intake and its role in reducing the risk of pancreatic cancer, given the importance of identifying risk factors. This systematic review followed the guidelines of the Cochrane Collaboration and the Meta-analysis of Observational Studies in Epidemiology and the Preferred Reporting Items for Systematic Reviews and Meta-Analyses 2020. The structured literature search was conducted on PubMed/Medline and Scopus, combining free text words and medical subject headings. Our review contained 18 records at the end of the process. Our results show that dietary fiber intake reduces the risk of pancreatic cancer. When the analysis was differentiated according to the type of fiber considered, sub-grouped by gender (reduction of around 60% among women), and when case-control studies were conducted, the strength of the association increased. Clinicians and policymakers should improve interventions to raise the population’s awareness regarding the consumption of high-fiber diets, both in practice and in terms of public health policy. Pancreatic cancer has one of the worst prognoses among cancers, with a high case fatality rate and an overall 5-year survival rate of around 5% ,2. AlthoCigarette smoking, alcohol consumption, chronic pancreatitis, obesity, and diabetes mellitus have all been identified as risk factors for this illness . In termFiber intake has been ascertained as an essential component of a healthy diet . The benGiven the considerable burden of pancreatic cancer and the importance of identifying risk factors that could be modified by healthy nutritional recommendations, we conducted a systematic review and meta-analysis with the goal of summarizing the current evidence on dietary fiber intake and its role in reducing the risk of pancreatic cancer.This systematic review follows the Cochrane Collaboration  and the Articles had to meet the following criteria to be considered eligible: (i) written in English; (ii) population: adults ≥18 years ; (iii) interventions and exposures: highest dietary fiber intake; (iv) comparators/control: lowest dietary fiber intake or no intake; (v) type of study: epidemiologic studies .Exclusion criteria were (i) articles not published in English; (ii) people under the age of 18; (iii) full text not available; (iv) interventions and exposures: fiber supplementation; (v) comparators/control: fiber supplementation; (vi) type of study: review article, meta-analysis, trial, expert opinion, commentary, or article with no quantitative information or details. A detailed description of inclusion/exclusion criteria is reported in Two review authors (OES and VG) independently screened titles and/or abstracts of studies collected using the search strategy and those gathered from additional sources to identify research that would meet the inclusion criteria outlined above. Two members of the review team (OES and VG) independently assessed the downloaded full text of these studies for their eligibility. Any disagreements about the eligibility of particular studies were resolved through discussion with a senior reviewer (MN). In line with previous studies ,18, a stA “flow diagram” charting the number of references at each stage in the review process was produced in line with the PRISMA 2020 guidelines . The quaTwo researchers (CF and PF) independently performed the critical appraisal using the Newcastle-Ottawa Scale (NOS) , a risk 2 and I2 tests were used to assess the heterogeneity of the included studies. Heterogeneity was defined as high when I2 values > 75%, moderate when I2 values were between 50% and 75%, low for values ranging between 25% and 50%, and no heterogeneity for values below 25%. The graphical evaluation of the Funnel plot and the Egger’s regression asymmetry test were used to estimate potential publication bias; statistical significance was set at p < 0.10 [® software .The effect size (ES) was calculated based on the odds ratio (OR), risk ratio (RR), hazard ratio (HR), and mean and sample size provided for each study. The ES was reported as OR with a 95% confidence interval. Where articles did not directly report the OR, but instead gave the number of events (cases) among those exposed and not exposed and the mean value for dietary fiber intake for each group, the ORs and CIs were calculated from these data and included in the meta-analysis. Subjects having the highest dietary fiber intake were compared to those with the lowest (or no) dietary fiber consumption. A meta-regression analysis was conducted in the case of homogeneity in the reporting of dietary fiber intake (in terms of unit of measure). We used both fixed and random models in the current meta-analysis. This approach was chosen because, in most cases, a fixed model is used when studies are determined to be similar. However, the random effect model is strongly recommended when heterogeneity is classified as moderate to high. Chip < 0.10 . A trim In the case of studies using the same cohort (or study population), a sensitivity analysis was performed considering only the study with the highest quality score (QS) or with a larger sample size where the QS was equal, to rule out the possibility of overlapping effects. Moreover, the meta-analysis excluded studies with a computed OR. Furthermore, the sensitivity analyses only included studies with a follow-up (FU) equal to or greater than 9 years, validated tools to assess dietary fiber intake, type of diagnosis, and a QS greater than 7 . A subgrA cumulative analysis is a sequential meta-analysis, intended to evaluate how adding one study at a time modifies the ES. We ran three cumulative analyses: the first adding studies chronologically ; the second based on increasing sample size ; and the third based on ascending dose of dietary fiber intake (from the lowest to the highest). These types of analyses improve the potential consistency of the results .n = 261), the articles were not original (n = 193), they were not written in English (n = 25), or were not based on human studies (in vitro (n = 65) and in vivo studies (n = 36)). The full text of the 26 records was downloaded, and eight records were excluded with reasons following an in-depth assessment. The electronic searches on PubMed/MEDLINE and Scopus identified 274 and 382 records, respectively, for a total of 656 records. After screening the reference lists, four records were added, while 54 records were removed due to duplication. A total of 606 records were then assessed for eligibility. After screening the title and abstract, 580 records were eliminated because the topic was unrelated (dies n = ). The fun = 7; two studies published in Finland [n = 2) [n = 1) [n = 1) [n = 1) [ Finland ,50, one  Finland , in Ital Finland , in Pola Finland , in the  Finland , and the Finland ); five s Finland ,43,48,52 [n = 2) ,42, Aust [n = 1) , China ( [n = 1) , and Jap [n = 1) . Finally [n = 1) .n = 15) [n = 10) matched cases and controls by sex and age, with three studies matching by residency as well [Only three records were cohort studies ,46,49, wn = 15) ,50,51,52n = 15) , one wasn = 15) , and one as well ,47,51, w as well ,48. More as well ,43,48,51n = 14) investigated dietary patterns from at least 1 year to 10 years before the pancreatic cancer diagnosis, regardless of the technique employed to assess dietary intake. Finally, due to the poor prognosis associated with the disease, the subjects’ proxies  were allowed to be interviewed in seven studies in order to estimate dietary fiber intake. In terms of diagnostic assessment, a histological/cytological or radiological confirmation was used in eight studies [n = 7) studies [n = 2) [The mean age of participants in virtually all of the included studies was 60 years old, with the exception of five studies in which the age of the participants was not reported ,42,45,48 studies ,45,51,52 studies ,46,48,51 [n = 2) ,50, and  [n = 2) , while i [n = 2) . Funds w [n = 2) ,38,43,49The quality of the 18 studies ranged from 4 to 9, with a median value of 7. The assessment revealed a slightly higher quality level for cohort studies than for case-controls, and an overall increasing trend over time. The selection items, which consider the representativeness of the studies’ participants, mostly achieved high-quality criteria across the reports. On the other hand, serious flaws were discovered in the assessment of outcome/exposure, as well as in possible controlled factors. Quality evaluation results are presented in p-value < 0.001; in the random effect model, pooled ES = 0.63 (95% CI = 0.53–0.76), p-value < 0.001; based on 343,120 subjects with a high statistical heterogeneity ) (p = 0.006), no publication bias was found in either the fixed or random effect models (p = 0.175). Results were not confirmed in the random effect model . There were no material changes in results when the analysis was restricted to those studies that used a validated dietary assessment tool, establishing a significant association between a higher intake of fiber and a lower risk of pancreatic cancer in the fixed and random effect models ,  0.001)) a. As seet models b. We elit models . We alsot models . As regat models . When cot models , with mot models . In the In both the fixed and random effect models, the subgroup analysis based on soluble and insoluble fiber individually found a significant association between a higher intake of dietary fiber and a lower risk of pancreatic cancer . HoweverThe findings of our systematic review and meta-analysis revealed that dietary fiber intake has a preventive effect against pancreatic cancer risk, which remained consistently significant across the sensitivity or subgroup analyses performed.Worth noting is that the strength of the association increased when the analysis was differentiated based on the type of fiber considered (soluble or insoluble), and subgrouped by gender , and study design (higher in case-control studies). However, results showed a weak inverse borderline association when the diagnosis relied on cancer registries, or when only cohort studies were considered. The strongest link between dietary fiber intake and pancreatic cancer was seen when only case-control studies were included rather than cohort studies, which is likely due to the small number of cohort studies found. Moreover, the period of the three cohort studies varied significantly (ranging from 10 to 30 years of follow-up). Cohort studies are also more prone to selection bias due to the lost-to-follow-up phenomenon, especially for extremely long studies. Even case-control studies, however, are prone to some limitations, such as recall bias, mainly because the difference between cohort and case-control studies is that in case-control studies, the exposure  has already occurred in the past. Similarly, while the use of population-based cancer registries is of utmost importance for cancer surveillance, certain considerations (such as reporting delays and gaps) may limit their use. In particular, as suggested by Izquierdo et al., some disadvantages have been acknowledged in regard to the assessment of risk factors that are less detectable over longer time periods due to recall problems and difficulties in obtaining medical records from the distant past . In termn = 4/18). This aspect may have influenced the quality of available data due to fewer diagnostic tools as well as lower reporting accuracy, given that most scientific report guidelines are relatively new.We also performed a sensitivity analysis stratified by type of fiber. Dietary fiber is defined in the literature as non-digestible carbohydrates plus lignin , which cFinally, public health recommendations and healthy dietary guidelines all recommend a higher consumption of dietary fiber, even if no specific recommendations are available with regard to dietary fiber intake and pancreatic cancer. In fact, the most recent updated version of the World Cancer Research Fund/American Institute for Cancer Research (WCRF/AICR) report, which is used as a reference for cancer prevention through diet and nutrition, found limited evidence on dietary fiber intake and pancreatic cancer risk .Nevertheless, the results of our systematic review and meta-analysis revealed that dietary fiber consumption protects against both prevalent and incident pancreatic cancer, also in line with literature so far available ,56. TherOur findings have important implications on clinical practice, since recommended dietary fiber intake as part of healthy dietary pattern is also beneficial in pancreatic cancer prevention. This is especially true when considering the general public’s generally limited adherence to healthy eating recommendations . CliniciThis study’s strengths and limitations should be acknowledged before generalizing the results of our systematic review with meta-analysis. To begin with, this is the most updated review assessing the association between dietary fiber intake and the risk of pancreatic cancer. In fact, Wang et al. published the first meta-analysis on the same topic in 2015, although only one cohort study and a lower number of case-control studies were obtained at that time . Moreoven = 54) was lower than expected, likely because of the type of journals in which papers were published that are not indexed on more than one medical archive, that can confirm a medium or low-quality score of the included reports. In some cases, certainty with regard to dietary intake assessment is low due to the method used, or because proxies were also considered as valid respondents. However, independently of the respondent type, all dietary intake estimations derived from self-reported consumption which, as discussed above, even if validated, may be affected by some bias . Since most of the retrieved studies were relatively old, it was not possible to perform a sensitivity analysis by pancreatic cancer subtype. Furthermore, some other limitations could be attributed to the review itself, as we only included articles published in English, and this might have affected the total amount of eligible studies. However, English is the commonly-accepted language within the scientific community and high-quality findings are usually published in international journals that only accept papers in the English language.The limitations should also be given due consideration. First of all, our results are affected by the intrinsic limitation of the original studies retrieved. As mentioned above, several of the retrieved studies have a medium or low-quality evaluation. It is worth also noting that the number of duplicates retrieved (To conclude, our results demonstrate the beneficial effect that dietary fiber intake has in reducing the risk of pancreatic cancer. This protective role is particularly evident when considering the prevalence of pancreatic cancer, rather than the incidence. However, this result may be due to the very low number of cohort studies retrieved. Future prospective cohort studies should be encouraged in light of this. The results of this meta-analysis reveal that dietary fiber can protect both women and men. Furthermore, the robustness of results increased when a subgroup analysis by type of fiber was performed. Finally, a dose meta-regression analysis was conducted, which confirmed a weak inverse borderline association between a higher intake of fiber and a lower risk of pancreatic cancer. This systematic review’s pooled analysis, as well as several sensitivity and subgroup analyses, indicated the consistent and robust beneficial effect of dietary fiber intake on lowering the risk of pancreatic cancer.


Objective and Methods. The PubMed, Embase, and Wanfang databases were searched, covering all publications before July 20, 2020.  Atrial fibrillation (AF) is a common, sustained cardiac arrhythmia. Recent studies have reported an association between ZFHX3/PRRX1 polymorphisms and AF. In this study, a meta-analysis was conducted to confirm these associations. P < 0.001). Similarly, strong associations were detected through stratified analysis using source of control and genotype methods . In contrast, an inverse relationship was observed between the PRRX1 rs3903239 polymorphism and AF risk . No obvious evidence of publication bias was observed.  Overall, seven articles including 3,674 cases and 8,990 healthy controls for ZFHX3 rs2106261 and 1045 cases and 1407 controls for PRRX1 rs3903239 were included. The odds ratio (OR) ) was used to assess the associations. Publication bias was calculated using Egger's and Begg's tests. We found that the ZFHX3 rs2106261 polymorphism increased AF risk in Asians (for example, allelic contrast: OR [95% CI]: 1.39 [1.31–1.47],  In summary, our study suggests that the ZFHX3 rs2106261 and PRRX1 rs3903239 polymorphisms are associated with AF risk, and larger case-controls must be carried out to confirm the abovementioned conclusions. Atrial fibrillation (AF) is a common form of arrhythmia, with an incidence of approximately 1% among adults worldwide . PreviouP=2.76 × 10−7). At the same time, Gudbjartsson et al.  = 1.57 [1.38–1.79], P(heterogeneity) = 0.156, P < 0.001), and PB  = 0.321, P < 0.001; AC vs. CC: OR [95% CI] = 1.17 [1.04–1.30], P(heterogeneity) = 0.584, P=0.007) , the heterozygote comparison , AA vs. CC , the dominant model , and AA vs. AC + CC , high-resolution melt , the heterozygote comparison , AA vs. CC , the dominant model , and AA vs. AC + CC , and others (data not shown)) ) .Pheterogeneity = 0.522, P=0.036, P=0.137 for heterogeneity, P=0.006, Decreased associations were found in the heterozygote comparison ; z = 1.2, P=0.23 (Begg's test) for ZFHX3 rs2106261, vs. T-allele, t = 0.11, P=0.933 (Egger's test); z = 0.0, P=1.00 (Begg's test) for PRRX1 rs3903239, Begg's funnel chart and Egger's test were performed to assess publication bias. The results did not show any evidence of publication bias  [A network of potential gene-gene interactions for ZFHX3 and PRRX1 genes was analyzed using the String online web page (db.org/)   [P=4.2 × 10−5 for rs3903239 and 3.87 × 10−6 for rs2106261) [Previously, some relative studies have been reported. Zaw et al. showed that ZFHX3 rs2106261 polymorphism was a risk marker for AF and AF-related phenotypes . Huang e× 10−11) . However× 10−11) . Okubo e2106261) . Liu et PRRX1 rs3903239 polymorphism and AF risk from a comprehensive analysis, involving 4719 cases and 10397 controls [The current analysis is to evaluate the associations between ZFHX3 rs2106261 or controls . We founβ1) (score = 0.900). Several studies have focused on CDKN1A and TGFβ1, but not RUNX3, in the development of AF. Further studies should focus on the three abovementioned potentially related genes and their common polymorphisms in AF. On the contrary, the scores of related genes for PRRX1 are generally low; however, this should be verified and indicated in future research.In addition, the online analysis system String was applied to predict the potential functional partners of the genes, which may help to expand the range of vision of related genes. Ten genes were identified. The three highest scores of associations were for cyclin-dependent kinase inhibitor 1A (CDKN1A) (score = 0.921), runt-related transcription factor 3 (RUNX3) (score = 0.918), and transforming growth factor-beta 1 (TGFP value was evaluation criteria to select the model for analysis, which may result in the final results. No publication bias was found, which may reduce the influence from the heterogeneity in our analysis.Although positive results were found, limitations of the current study should also be discussed. First, the literature published is relatively new, so the number of included studies is not sufficiently large based on current publications and more well-designed and larger studies in future research should be paid attention to. Second, it is possible that specific environmental and lifestyle factors influence the associations between ZFHX3 rs2106261 or PRRX1 rs3903239 polymorphism and AF including family history, age, sex, disease stage, and lifestyle. Moreover, whether the AF patients have other complications, such as hypertension, diabetes, and coronary heart disease, all the included papers have not been reported. Further comprehensive studies should include the abovementioned information. Third, there are several types of AF, such as persistent, permanent, pathologic, idiopathic, and paroxysmal. If enough data exist for different types of AF in the future, we could classify the analysis into subgroups prior to analyzing the association of the ZFHX3 rs2106261 or PRRX1 rs3903239 polymorphism with AF, which could offer more precise findings for faster translation to the clinic. Fourth, the heterogeneity was existed in our analysis, such as in total and genotype method subgroup for rs2106261 and in total for rs3903239 polymorphism. The heterogeneity for Our analysis illustrated that the ZFHX3 rs2106261 and PRRX1 rs3903239 polymorphisms are associated with conspicuous AF risk in Asians. Therefore, well-designed and larger studies, including information about gene-gene/gene-environment interactions, are recommended to confirm the abovementioned conclusions.


Periodontal disease is a chronic infectious disease caused by bacterial infection which may lead to various systematic diseases. Recently, increasing studies have explored the correlation of periodontal disease with the risk of prostate cancer. However, the findings were inconsistent. Hence, this study aims to investigate the association between periodontal disease and the risk of prostate cancer by a meta-analysis. PubMed, EMBASE, and Cochrane were searched for publications up to July 17, 2020. Cohort and case-control studies evaluating the risk of prostate cancer in patients with periodontal disease were included. A fixed or random-effect model was used to calculate the summary relative risk (RR) along with 95% confidence interval (CI). All analyses were conducted using Stata 12.0 software.P = 0.001). Findings of sensitivity analyses proved that the overall results were robust. Seven studies were included in the final analysis. The pooled estimates showed that periodontal disease was significantly associated with the risk of prostate cancer (RR = 1.17; 95% CI = 1.07-1.27;  Periodontal disease may be considered as a potential risk factor for prostate cancer. Although it’s a possibility, males should be more aware of their oral health and implement effective measures to prevent and treat periodontal disease. Key words:Periodontal disease, periodontitis, prostate cancer, meta-analysis Prostate cancer (PC) composes the largest part of the diseases that affect men’s health worldwide, and its incidence increases with age . It is cet al.  is a chronic infectious disease caused by bacterial infection which invades gingiva and periodontal supporting tissues . It affeAt present, although some studies have found a close relationship between PD and PC -21, the This meta-analysis was based on the Preferred Reporting Items for Systematic Review and Meta-analysis (PRISMA) guidelines .- Eligibility criteriaStudies meeting the following criteria were included: 1) cohort study or case-control study; 2) the exposure of interest was PD, and the comparison was an absence of PD; 3) the outcome of interest was patients with PC; 4) the hazard ratio (HRs), relative risk (RRs) or odds ratio (ORs) estimates with their 95% confidence intervals (CIs) were reported; 5) the study with higher quality or better comprehensiveness was included if two or more articles used the same dataset; 6) studies were published in English.- Search strategyA comprehensive literature search was conducted with PubMed, EMBASE, and Cochrane to identify studies up to July 17, 2020, using the following search terms: (“prostate neoplasm” OR “prostate cancer” OR “prostate tumor” OR “prostatic cancer” OR “prostatic neoplasm” OR “prostatic tumor”) AND . No data and language restrictions were applied in searching. Additional studies were retrieved by searching reference lists of relevant articles.- Study selection and data extractionDuplicate records were removed, then study titles and abstracts were screened to make sure they were relevant. Subsequently, studies that satisfied the eligibility criteria were included through full-text assessment.The following data were obtained from each included study: first author’s surname, year of publication, country of origin, study design, sample size, assessment methods of exposure and outcome, age, follow-up period, the estimate  with 95% confidence interval (CI) and adjustment confounding factors. These data were independently extracted by two authors (YW and YZ). Any disagreements between these authors were resolved through discussion or by consulting a third author (RH).- Quality assessmentThe methodological quality of the included studies was assessed by Newcastle-Ottawa Scale (NOS) . The sca- Statistical analysisP < 0.1) and the I2 statistic (I2 ≥ 50% indicated significant heterogeneity). In the presence of significant heterogeneity, the random effect model was adopted; otherwise, the fixed model was used. Subgroup analyses stratified by study design, country of origin, follow-up period, sample size and quality of studies were conducted. Sensitivity analysis was performed to examine the robustness of the results by using a leave-one-out approach. Begg’s test and Egger’s test were used to assess the Publication bias. A P-value of less than 0.05 was regarded statistically significant. All analyses were conducted with Stata 12.0 software .RR with 95% CI was used as the pooled estimate to assess the association of PD and the risk of PC. When the outcome of interest was rare, it was considered that the OR approximated the RR . HR was - Study selection and characteristicet al. , exposure of interest (n = 2), complete data (n = 2), eligible comparison (n = 1) nor overlaps in datasets (n = 2) Fig. ,18,20,21et al.  which wa- Overall result and subgroup analysisP = 0.001). No significant heterogeneity across studies was observed  and four prospective cohort studies . For analyses based on the country of origin, the association of PD and the risk of PC was statistically significant among Asian populations  or European populations  but not among American populations . In terms of the duration of follow-up, the pooled results of four studies which followed target populations of < 14 years showed a significant relation between PD and the risk of PC ; no evidence of increased risk of PC among patients with PD was found in the studies with a follow-up period of ≥ 14 years . Analyses according to sample size revealed that PD was associated with the risk of PC in the studies with less than 10,000 participants  but not in the studies with more than 10,000 participants . Moreover, after the exclusion of the study which was not of high quality, the results nevertheless indicated a significant association between PD and the risk of PC .- Sensitivity analysisWe deleted a study from the overall meta-analysis each time to explore the influence of the removed data on the pooled RR. The results showed omission of any individual studies brought no significant changes on pooled RR with 95% CI, indicating the meta-analysis results were robust .- Publication bias analysisP = 0.230) and Egger’s test (P = 0.078) indicated there was no statistically significant publication bias across those studies.The results of Begg’s test , used to diagnose PC, was significantly higher in patients with PD (et al. found that oral pathological bacteria are able to up-regulate or down-regulate pro-inflammatory cytokines and chemokines which affect oral and systemic immune system of the body, causing carcinogenic reactions (gingivalisPorphyromonas  and Actinobacillus actinomycetemcomitans were proven to be associated with a high risk of pancreatic cancer (et al. (At present, certain underlying mechanisms suggest that there may be a causal relationship between PD and PC. Researchers generally believe that PD causes cancer mainly by affecting systemic immune response which leads to the increase of serum inflammatory markers, and moreover, the level of C-reactive protein particularly increases with the progress of PD . Also, J with PD . Some st with PD ,27. Meurc cancer . However (et al.  have repThe meta-analysis performed by the authors of this paper consists of the following attributes: First, low heterogeneity across studies was observed, and there was no publication bias in the analyses. Second, the majority of studies included in the meta-analysis were considered to be of high quality according to the NOS. Although one of the studies was of moderate quality, no evidence of unsTable results was observed in the sensitivity analysis when excluded. Third, the selection and recall bias were minimized since all studies in our analyses were cohort studies. Fourth, as most human cancers show a long latent period, it is necessary to have a long duration of follow-up designed for studies. In this meta-analysis, most studies had more than 10 follow-up duration, and three of them were with a follow-up period of more than 14 years.Despite these strengths, there were several limitations of this meta-analysis. First, according to the results of subgroup analysis, the estimates from the subset of studies on target populations from the United States with longer follow-up or smaller sample size indicated that PD had no significant association with the risk of PC, which meant several factors may affect the correlation between PD and the risk of PC. The interpretation was yet unconvincing due to the limited number of studies in each subgroup. Second, although the majority of confounding factors were adjusted in the included studies, few studies adjusted for dental health status, stress and occupation, which may influence the precision of analyses. Third, it is clear that familial and genetic risk factors lead to the increased risk of PC , howeverIn summary, the present meta-analysis revealed that PD is significantly associated with the risk of PC, and that PD may increase the risk of PC. Based on the results, the general public should be more aware of their periodontal health and maintain a good oral health which may be beneficial in reducing possible risk of PC. However, well-designed studies with a more extended follow-up period, better assessment methods of PD, populations from diverse geographic regions and solid multivariable adjustments are warranted to confirm our findings.


We aimed to map and describe the current state of Mendelian randomization (MR) literature on cancer risk and to identify associations supported by robust evidence.robust, probable, suggestive, and insufficient, based on the significance and concordance of the main MR analysis results and at least one of the MR-Egger, weighed median, MRPRESSO, and multivariable MR analyses. Associations not presenting any of the aforementioned sensitivity analyses were not graded.We searched PubMed and Scopus up to 06/10/2020 for MR studies investigating the association of any genetically predicted risk factor with cancer risk. We categorized the reported associations based on a priori designed levels of evidence supporting a causal association into four categories, namely robust, 275 (18.7%) by probable, and 89 (6.1%) by suggestive evidence. The most prominent robust associations were observed for anthropometric indices with risk of breast, kidney, and endometrial cancers; circulating telomere length with risk of kidney, lung, osteosarcoma, skin, thyroid, and hematological cancers; sex steroid hormones and risk of breast and endometrial cancer; and lipids with risk of breast, endometrial, and ovarian cancer.We included 190 publications reporting on 4667 MR analyses. Most analyses  were not accompanied by any of the assessed sensitivity analyses. Of the 1467 evaluable analyses, 87 (5.9%) were supported by Despite the large amount of research on genetically predicted risk factors for cancer risk, limited associations are supported by robust evidence for causality. Most associations did not present a MR sensitivity analysis and were thus non-evaluable. Future research should focus on more thorough assessment of sensitivity MR analyses and on more transparent reporting.The online version contains supplementary material available at 10.1186/s12916-022-02246-y. With a global burden of 18.1 million new cases and 9.9 million deaths in 2020  ligand 1 and chemokine [C-C motif] ligand 4) with a median of 2.9%  reported a numerical estimation of the power of the MR analysis, with a median reported power of 76% (range 1 to 100%)  adjusted for population stratification, 907 (68.4%) adjusted for age, 720 (54.3%) for sex, and 271 (20.4%) used adjustments specific to genotyping methods. Other adjustments included study location or assessment center , anthropometrics , lifestyle factors , and study year/time , whereas in 81 (1.7%) analyses a number of additional adjustment factors were used.The median number of SNPs used as instruments was five, ranging from one to 3163, whereas for 141 (3%) MR analyses this information was not reported  and only 363 (7.8%) used a one-sample design. The statistical analysis method of preference as main analysis with 2974 (63.7%) associations was the inverse-variance weighted method (either fixed-effect or random-effects), whereas 734 (15.7%) associations were derived from likelihood-based analyses. Other statistical analysis approaches used for the main MR analysis included the Wald ratio, generalized models , two-stage regression approaches (35% of the one-sample designs), WM, and MR using robust-adjusted profile scores. Forty-two publications (22.1%) performed an adjustment for multiple comparisons, and from the 4667 total associations only 523 (11.2%) were statistically significant in the main analysis at the threshold set up by the study due to multiple testing or at nominal significance  if no multiple testing threshold was defined. Sensitivity analyses were mostly performed in two-sample MR, and a limited number of these sensitivity analyses were performed in one-sample MR designs.Most analyses were based on a two-sample  and 171  associations, respectively , 27 a WM , no analysis performed MRPRESSO, and 27 performed a MVMR analysis   analyses with 140 (10.8%) of those presenting a nominally statistically significant MR-Egger slope; a total of 1055 (24.5%) associations performed a WM analysis with 217 (20.6%) being statistically significant, while sensitivity analyses using MRPRESSO or multivariable MR were fairly limited with only 142  across 123 publications only presented results for the main analysis and therefore could not be graded. Of those 3200 associations, 293 (9.2%) had a one-sample and 2907  a two-sample design. For 36.6% (N = 1171) of analyses, the authors selected the IVs based on their biological relevance to the exposure, with 1106 (94.5%) of them having a two-sample design. A total of 238 (7.4%) associations with only a main analysis were statistically significant (or survived a multiple testing threshold) and for only 60 (25.2%) of those the selection of the instrument was based on biological relevance. Of those, 14 used a cis definition for the selected instruments, but none of those performed a colocalization analysis.A total of 1467 (31.4%) MR associations reported in 121 publications presented results on both the main and at least one sensitivity analysis and were further evaluated based on the aforementioned grading scheme. The rest of the MR associations , 275 with probable evidence , 89 with suggestive evidence , and 1016 with insufficient evidence  based on the results of the main and sensitivity analyses. Across the 16 exposure categories, anthropometrics had the highest number of robust analyses , followed by steroids , circulating leukocyte telomere length , the other diseases and traits category , and lipids , whereas no robust association was found among the amino acids and derivatives, fatty acids and derivatives, inflammatory biomarkers, methylations, and other metabolites and biomarkers categories (Table robust associations was observed for breast cancer with 29 (33.3%) of the 87 robust associations, followed by lung cancer  and endometrial . Head and neck, stomach, small intestine, pancreatic, cervical, and central nervous system cancers did not present any robust MR associations  and waist-to-hip ratio (WHR) with decreased risk of total breast cancer [robust associations were observed in the steroids category, pertaining to the positive association of different measures of testosterone with breast  and endometrial cancer, and to the negative association of sex-hormone-binding globulin (SHBG) and endometrial cancer [robust associations were also found for longer (shorter) leukocyte telomere length pertaining to increased  risk of total cancer [robust associations from the lipid metabolism biomarkers category pertained to high-density lipoprotein cholesterol (HDL-C) with increased risk of breast  [robust associations pertaining to magnesium with breast  [robust associations pertained to age at menarche with ovarian  [The 16 t cancer , 299, est cancer , 299, ant cancer , 299); Bt cancer  and endot cancer  cancer, t cancer  and ovart cancer . Thirteel cancer . Thirteel cancer , lung   but decrR+, ER−) ; triglycR+, ER−) ; low-denR+, ER−)  and lungR+, ER−) ; total cR+, ER−) ; and 3-hR+, ER−) . From thR+, ER−) , 328, SCR+, ER−)  and one R+, ER−) . From thed risk) , ferritied risk) , alcoholed risk) , and vited risk) . Transfeed risk) . The resed risk) , alcoholed risk) , endometed risk)  and withed risk)  (both ined risk) , insulined risk) , obstruced risk) , polycysed risk) , stem ceed risk) , schizoped risk) , standared risk) , thyroided risk) , type 2 ed risk) , and vited risk) .probable and four with suggestive evidence were upgraded to robust, while 35 associations were upgraded from suggestive to probable. In contrast, 23 MR analyses with probable and 32 with suggestive evidence were downgraded to insufficient evidence. Finally, 15 associations with robust evidence, 34 with probable, 17 with suggestive, and 242 with insufficient evidence now presented only a main analysis and were non-evaluable , despite most studies being conducted in an era of large GWASs across a wide breadth of phenotypes. This may partially be explained by the large number of infrequently used biomarkers that were assessed in some studies [r2 < 0.001 to r2 < 0.1, one in ten had an even more liberal threshold. Researchers should consider adjusting for the potential correlation between IVs when using less strict thresholds such as 0.1 or higher [R2 and a priori power estimation are equally important for evaluating the capacity of an IV to provide valid and accurate estimates and can help to differentiate between non-significant but otherwise underpowered associations from real null ones.The median number of IV size across all analyses was relatively small , which was supported by robust evidence across several different MR analyses. In contrast, observational evidence supports a positive association of body fatness with post-menopausal breast cancer risk, and an inverse association for premenopausal disease [Robust evidence was also observed for a positive association of BMI and endometrial cancer in Asian populations [robust evidence for BMI and probable evidence for WHR and body fat percentage, both of which were upgraded to robust in the sensitivity analysis excluding the MR-Egger analysis. Several well-acknowledged observational associations of adiposity and cancer risk, namely for ovarian [probable evidence. The association for ovarian cancer from the largest MR study to-date failed to reach robust evidence due to the main analysis not surviving the multiple comparisons threshold set by the original publication that investigated many risk factors, despite being nominally significant [robust evidence due to several reasons, including not surviving the multiple correction thresholds and having non-significant sensitivity analyses. BMI also presented probable evidence of an increased risk with lung SCC. The results from observational data are showing inverse associations for BMI and risk of total lung cancer [robust evidence.Across the MR analyses pertaining to anthropometric exposures,  disease , 331. Th disease , 332, an disease  and in M disease , 299 witulations , which iulations , 335. Thulations , 236, soulations , 337 was ovarian , 338 and ovarian , 339 cannificant . Similarnificant , 167, dig cancer , 340, whg cancer . With reg cancer  and ovarg cancer  that reaRobust and probable evidence was also found for the positive association of genetically predicted testosterone concentrations with risk of breast and endometrial cancer, and the negative association of SHBG with endometrial cancer. These results have been partially confirmed in observational evidence [evidence , 343. Coevidence  and endoevidence . On the evidence .robust associations were observed with total cancer, and with lung, leukemia, lymphoma, osteosarcoma, skin, and thyroid cancers, where longer telomeres increased the risk (or shorter lengths decreased the risk) of these cancers. Furthermore, a positive association with increased telomere length was supported by probable evidence for a number of other cancer sites, such as glioma, bladder, kidney, melanoma, multiple myeloma, non-Hodgkin’s lymphoma, ovarian, and prostate cancer, several of which were upgraded to robust with the exclusion of the MR-Egger analysis. In contrast, negative associations of increased telomere length with cervical, head and neck, pancreatic, and skin basal cell cancers were supported by probable evidence. The observational evidence has created controversy in the literature about the direction of the associations [TERT, TERC, and OBFC1 [TERT region and lung cancer risk is mediated by telomere length [A considerable fraction of the studies focused on circulating leukocyte telomere length, for which ciations , 348, whciations . A recennd OBFC1 , while me length . The cure length , 351, ale length .robust and probable evidence, which is in line with the observational evidence [robust evidence for a negative association with LDL-C and lower levels of evidence for associations with other lipids [robust association was observed between HMG-CoA reductase, the drug target of statins, and lower risk of ovarian cancer with consistent MVMR results accounting for BMI. Observational evidence for statin use suggests a decreased risk of ovarian cancer among statin users [robust evidence with lung SCC, pertaining to a negative association for total cholesterol and LDL-C, but MVMR analyses were not conducted, while for total lung cancer these associations were supported by probable evidence. Observational studies indicated a lower risk of lung cancer for circulating lipids [robust evidence. The role of lipid metabolism in carcinogenesis and tumor growth has been acknowledged in the literature [Several associations were identified for lipids, especially TGL, total cholesterol, LDL-C, and HDL-C. Specifically, the negative association of TGL with total and ER+ breast cancer was supported by evidence , 354. Foevidence . With rer lipids . These rr lipids –358 but r lipids , 360. Anin users . Only twg lipids . For sevterature , 364 altterature , 366. Reterature . An overterature .Many of the included associations were non-evaluable due to not performing any of the sensitivity analyses required for our grading. Reasons may vary across studies, including inability to do so due to low number of instruments, especially for the MR-Egger analyses, prioritization of statistically significant associations for further evaluation with sensitivity analyses, or sensitivity analyses not being part of the authors’ analysis plan. There is a necessity to study these associations more comprehensively, especially in the cases of polygenic definition of instruments, which are more prone to biases or pleiotropy that can drive associations both towards and away from the null. Regardless of the reason and the appropriateness of the decision to include sensitivity MR analyses, these associations are not sufficiently investigated and are all considered non-evaluable in our grading scheme, which focuses on evaluating the robustness for causality of the studied associations.Other efforts to summarize the evidence of MR analyses on cancer risk have been performed previously. However, they were either limited to specific exposures , 18, 20 non-evaluable evidence. The structure of the criteria for evaluation of the robustness of the MR evidence for causality was more geared towards the evaluation of two-sample MR approaches, but the percentage of one-sample designs that did not perform one of the pre-specified sensitivity analyses was only marginally higher than that of two-sample designs. Associations evaluated in earlier publications, especially those before many of the sensitivity analyses were introduced, could also not be evaluated. However, the majority of the studies were published after 2018 and the earlier associations often relied on limited number of cases or on instruments including only a limited number of SNPs and with low percentage of variability explained. Information of the percentage of variance explained and statistical power of the instrument was often not reported, and thus a complete assessment of weak instrument bias could not be performed. Therefore, the grading scheme did not allow us to distinguish MR analyses that presented robust evidence of lack of association from MR analyses that did not present an association due to being insufficiently powered. Future studies may benefit from reporting this information. The approach undertaken in this review for grading the associations did not allow to us to evaluate MR analyses that only presented a main analysis without being supported by sensitivity analyses. Since two of the three MR assumptions are not directly testable, a MR analysis is imperative to be supported by a comprehensive evaluation of complementary and sensitivity analyses to increase credibility of the results, as such approaches can at least give some indication of large violations of the assumptions. Most MR analyses evaluating associations for gene products using cis instruments were non-evaluable using our current criteria as most included one or two SNPs as IVs, and the sensitivity analyses could not be applied. However, only two of these studies performed colocalization analysis and neither presented statistically significant associations for these specific analyses. More recently introduced sensitivity MR analyses were not included in the current evaluation, as their use is very infrequent in the MR literature. Finally, there is discrepancy in the availability of genetic data for different cancers, and hence the MR studies that have been possible; thus, cancer consortia are encouraged to make their summary data more readily and widely available.Several limitations need to be acknowledged. Our search strategy may have resulted in missing some relevant studies, especially if the MR analysis was not the primary focus of some studies but only a supplementary analysis, which seems to be increasingly common in recent GWA studies. In these cases, however, we would not expect a comprehensive evaluation of the studied associations using sensitivity MR analyses, which would only lead to inflation of the number of associations with robust evidence. In addition, we identified gaps in the conduct and reporting of MR studies that will assist in developing stronger future reporting guidelines.The field of cancer epidemiology is challenging to evaluate due to the sheer amount of available observational evidence and further burdened by the increasing interest on MR methodologies that could complement findings from traditional observational research. Our work summarizes and evaluates the robustness of the MR analyses evidence for causality in cancer prevention and etiology. Only a minority of the evaluated MR analyses were supported by Additional file 1. Supplementary methods.Additional file 2: File S1. List of excluded studies and reason for exclusion. File S2. List of distinct exposures per exposure category. File S3. Evidence base. Table S1. Description of instrument characteristics by exposure category. Table S2. Summary results of the main and sensitivity Mendelian randomization (MR) analyses by exposure category and cancer group. Table S3. Grading of the evidence based on the sensitivity analysis by excluding the results from the MR-Egger test.


Helicobacter pylori (H. pylori) infection and nonalcoholic fatty liver disease (NAFLD) is a matter of debate. Although it has been studied in many observational studies, the results remain controversial. Therefore, we performed a meta-analysis to assess the association between H pylori infection and risk of NAFLD.The relationship between 2-index), subgroup analyses, regression analyses, sensitivity analysis and the possibility of publication bias were assessed.We searched Pubmed, EMBASE, and Web of Science databases, from inception to September 10, 2020. Odds ratio (OR) and 95% confidence interval (CI) were pooled by random-effects model. The statistical heterogeneity among studies . The results of meta-regression implicated that the study type and the case-control ratio impacted the total effect size. Funnel plot did not show significant publication bias. Meta-analysis of data from longitudinal studies showed that H pylori infection was also associated with increased NAFLD incidence .A total of seventeen studies involving 91,958 individuals were included in our meta-analysis. Meta-analysis of data from cross-sectional and case-control studies showed that H pylori infection and the risk of NAFLD. Further studies are required to strengthen the association and clarify the mechanism.The results indicated that a positive association between  The human gastrointestinal tract is an integrated ecosystem, known as microbiota. Mucosal epithelium of the small intestine is a barrier between gut lumen and microbiota. Intestinal microflora is the primary source of endotoxins produced by Gram-negative bacteria (lipopolysaccharide), that normally cross the mucosa in trace amounts, enter the portal circulation and finally clear to liver cells. The increased intestinal permeability appears to be caused by disruption of the intercellular tight junctions. Helicobacter pylori (H. pylori) is a Gram-negative, spiral-shaped bacterium that colonizes the gastric epithelium. More than half of the world's population is infected with H pylori. Recent studies have shown that, H pylori infection plays a potential role in insulin resistance and increased intestinal permeability, which may contribute to the development of NAFLD. These studies have prompted our interest in updating current evidence. Hence, we performed an updated meta-analysis of all available evidence published to date regarding the relationship between H pylori infection and NAFLD.Nonalcoholic fatty liver disease (NAFLD) is characterized by hepatic steatosis while excluding alcohol use and other causes of liver diseases which is the leading cause of chronic liver diseases in the world.22.1This meta-analysis was conducted following the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines, and the protocol for this meta-analysis is available on international prospective register of systematic reviews .2.2Helicobacter pylori” OR “campylobacter pylori” OR “H pylori” OR “HP” OR “Helicobacter spp” OR “H. pylori”) AND . Both Medical Subject Heading and free word were used. We also reviewed references from relevant original papers and review articles to identify further eligible studies not covered by the original database research. This review was performed according to the guidelines for meta-analyses and systematic reviews of observational studies.Pubmed, EMBASE, and Web of Science databases were electronically searched from inception to September 10,2020, without language restrictions. The search terms were as follows:, risk ratio (RR), hazard ratios (HR) with 95% confidence intervals (95% CI) should have been provided;3.H pylori infection had to be confirmed at least one positive test as follows: serological testing (using H pylori IgG enzyme-linked immunosorbent assays), 13C-labeled or 14C-labeled urea breath test or fecal antigen test.4.NAFLD had to be diagnosed by histology, imaging (mostly ultrasonography) or surrogate markers of NAFLD, such as either the hepatic steatosis index (HIS) or the NAFLD liver fat score (NAFLD-LFS), which include in their equations serum transaminase levels, body mass index (BMI), fasting insulin levels, pre-existing diabetes or metabolic syndrome.5.all the studies included a control group.Criteria for exclusion were as follows:1.letters, abstracts, case reports, animal studies, editorials, reviews and meta-analyses;2.irrelevant literatures and duplicate studies;3.studies which did not exclude individuals with significant alcohol intake and other secondary causes of chronic liver disease;4.studies that did not specifically report any OR or HR and 95% confidence intervals (CI) for the outcome measure of interest.Two investigators (Lin Wei and Hui-Guo Ding) independently screened the titles and abstracts of all studies identified using the previously described search criteria to identify studies meeting the inclusion criteria. Each study meeting the requirements of the inclusion criteria then underwent a full-text independent review by both investigators. Disagreements about the inclusion of studies between investigators were resolved by discussion.2.4We extracted the following data from each study:1.study characteristics, including the name of the first author, publication year, country of publication, study design, sample size, mean age;2.H pylori infections in the NAFLD group, method of detection of both NAFLD and H pylori infection, the number of positive/negative H pylori infections in the control group, case/control ratio;the number of positive/negative 3.list of covariates adjusted in multivariable regression analyses.We assessed the quality of each study according to the Agency for Healthcare Research and Quality (AHRQ) and the Newcastle–Ottawa Quality Assessment Scale (NOS).2.5H pylori infection occurring in NAFLD patients versus the controls. Heterogeneity was assessed by Chi-square-based Q test and I2 -index were used to evaluate the statistical heterogeneity between the studies. The significance for the Q test was defined as P-value < .1. Heterogeneity was classified as follows: a I2 value of 0–25% indicated no heterogeneity, 26–50% indicated low heterogeneity, 51–75% indicated moderate heterogeneity, and 76–100% indicated high heterogeneity. a fixed effect model was used when I2 values of < 50%, and a random effect model was used when I2 values of > 50%. The subgroup analysis and regression analysis were performed to explore sources of heterogeneity. The forest plot assessed the relationship between H pylori infection and NAFLD. The funnel plot and Begg's and Egger's tests were used to investigate publication bias, P < .05 was considered statistically significant. What's more, we used the trim-and-fill method to further examine the possibility of publication bias.We used STATA version 12.0 software  to perform meta-analyses. Odds ratio (OR) with 95% confidence interval (95%CI) was pooled by this software, which was used to describe the ratio of 33.1The initial database search identified 104 records. Another record was identified from relevant studies. After removing 32 duplicates, 73 records remained. Of these 73 studies, 56 were excluded. Thus, 17 studies were included in the meta-analysis. A flow diagram of the literature search and the PRISMA flow diagram of the meta-analysis is shown in Fig. –21 1 case-control study, and 2 longitudinal studies.,24 11 of these studies were from Asia,,18,22,24 2 from the United States,,25 1 from Greece, 2 from Egypt,,23 and 1 from Switzerland.The main characteristics of the 17 studies are summarized in Table The results of quality assessment according to the Agency for Healthcare Research and Quality (AHRQ) for cross-sectional studies were presented in Fig. 3.2H pylori infection on the risk of prevalent NAFLD in cross-sectional or case-control studies  were shown in Fig. H pylori infection in NAFLD patients compared with the controls was 1.38 . Given the high heterogeneity, we carried out subgroup and meta-regression analyses according to the publication year, study type, the diagnosis of H pylori infection, the diagnosis of NAFLD, sample size, area, and case-control ratio. The result of subgroup and meta-regression analyses are shown in Table 2 = 84.55%, P = .001,χ2 = 1.2%] and  . We used a funnel plot to qualitatively detect the publication bias, an Egger's and Begg's tests to quantify the publication bias .The forest plot and pooled estimates of the effect of 44.1H pylori infection was associated with the infection risk of NAFLD . The results of meta-regression implicated that the study type and the case-control ratio impacted the total effect size. Meta-analysis of data from longitudinal studies showed that H pylori infection was also associated with increased NAFLD incidence .Our meta-analysis includes 17 observational studies. There are 14 cross-sectional studies,1 case-control study, 2 longitudinal studies. Meta-analysis of data from cross-sectional and case-control studies has shown that 4.2H pylori infection . Nevertheless, the search time was from inception to June 2016, and only six studies  were included in the study. Furthermore, the previous study did not provide a detailed subgroup analysis to explore the potential influence factors in NAFLD. In addition, numerous new relevant high-quality studies on the association between H pylori infection and NAFLD have emerged since that meta-analysis was published. Therefore, it is necessary to implement a new meta-analysis on this issue. In the last two years, Zhou et al published a meta-analysis of 15 observational studies  in 2019. Alessandro et al published a meta-analysis of 13 observational studies  in 2019. Compared to Zhou's and Alessandro's meta-analyses, our meta-analysis did not include congress abstracts.–29 As article published by Sumida et al did not contain control group, our meta-analysis also did not contain their data. Studies in our meta-analysis almost included the articles of Zhou's and Alessandro's. What's more, we supplemented 5 new studies published in 2019 and 2020.,15,19–21 In addition, we consulted relevant documents of the quality assessment, and we used the Agency for Healthcare Research and Quality (AHRQ) for the quality assessment of cross-sectional studies. The other two longitudinal articles were assessed by the Newcastle–Ottawa Scale (NOS).Recently, Wijarnpreecha et al conducted an excellent meta-analysis and demonstrated a significantly increased risk of NAFLD among patients with 4.3,33H pylori infection has been implicated in the pathogenesis of IR (insulin resistance) by many mechanisms, in particular: increased level of pro-inflammatory cytokines, eicosanoids, acute phase proteins, reactive oxygen species production, and cytokine serum changes. Many studies have demonstrated that H pylori infection could participate in the development of NAFLD through interfering with lipid metabolism and insulin sensitivity, stimulating and promoting oxidative stress and inflammation.Firstly, NAFLD is the hepatic component of metabolic syndrome and IR (Insulin resistance) is regarded as its key pathogenetic hallmark. The accumulation of liver fat in hepatocytes caused by insulin resistance makes the liver more vulnerable to oxidative stress and subsequent lipid peroxidation. These factors lead to the development of NAFLD.–42 In addition, H pylori presents a direct action on the hepatobiliary tract by toxins and its constituents, circulating in the portal blood; Patients with NAFLD present a significant increase in gut permeability, and this data is positively related with liver fat accumulation. H pylori invasion into intestinal mucosa might increase gut permeability, resulting in more passage of endotoxins via the portal vein to the liver.H pylori itself is also found in the liver and may cause inflammation and ballooning of hepatocytes. This H pylori-related hepatic injury may also play a role in the development of NAFLD.Secondly, Hp presence has been implicated in a variety of extra-digestive conditions, in particular vascular disorders caused by the release of vasoactive and pro-inflammatory cytokines.H pylori infection affect mainly aged population in the world. NAFLD is the most common liver disease in the world. A study published by Abd-Elsalam et al in 2017 showed that NAFLD was found in 42.5% of acute ischemic stroke patients. NAFLD might be associated with more severe stroke and worse outcome and the prevalence of diabetes was significantly higher among patients with NAFLD than those without NAFLD. On the other hand, ongoing persistence of obesity with increasing rate of diabetes will increase the prevalence of NAFLD, and as this population ages, many will develop cirrhosis and end-stage liver disease. a worldwide increasing prevalence of H pylori infection with age, reaching 40%–60% in asymptomatic elderly individuals and > 70% in elderly patients with gastroduodenal diseases.Thirdly, NAFLD and 4.4H pylori infection and the risk of liver cirrhosis. Second, our study was registered in advance on PROSPERO platform, and most included studies were of high quality, indicating our results were reliable. Third, we conducted subgroup analyses, meta-regression and sensitivity analyses to further evaluate the association, which made our results more credible, and understood the relationship more comprehensively.Our study has several strengths worth noting. First, we performed a rigorous search strategy and strict inclusion criteria, including all available evidence published to date. To our knowledge, ours is the largest and most updated meta-analysis to date aimed at investigating the association between H pylori infection was based on serological test (ELISA), rapid urease test (RUT), urea breath test (UBT), histology or multiple means. In all of these studies, serology is the most used diagnostic method, however, its reliability is lower than other diagnostic tools such as histology or RUT.Despite these strengths, some limitations of our meta-analysis should be acknowledged. First and foremost, most of original studies are retrospective or cross-sectional design, which can only at best demonstrate an association but not causality. And the results of observational studies are generally more susceptible to bias and confounding factors than randomized studies. So further large-scale prospective studies verifying the causal relationship are needed. Second, there is a high heterogeneity in the overall results. Heterogeneity may reduce the reliability of our conclusions. However, we conducted numerous subgroup and meta-regression analyses with the hope of detecting potential factors for such heterogeneities, to find the study type and the case-control ratio may be the source of statistical heterogeneity. And most results of subgroup analyses were consistent with overall results, indicating the reliability of our results. Third, the risk factors for NAFLD include dyslipidemia, obesity, age, environment, diet and sex, and additional biochemical features of selected articles cannot be extracted, so our data integration for the meta-analysis has not been able to control these factors. Several studies did not report OR in the report OR in the results and we could not calculate adjusted OR from all the included studies. Thus, the pooled overall OR was not adjusted for confounding factors such as age, sex, BMI, and diabetes. Fourth, most of the studies were done in the Asian area, so this result was more suitable for the Asian people. Fifth, there is no uniform standard for diagnosis of NAFLD in these studies. A majority of original studies used ultrasonography and fibro-scan to diagnose NAFLD. However, diagnosis by ultrasonography has inevitable limitations due to an incorrect diagnosis of NAFLD. The last, the diagnosis of 5H pylori infection and the risk of NAFLD. Further prospective studies are warranted to strengthen the association and to clarify whether there is a causative link between them. If this association is confirmed in the near future, H pylori infection eradication may be a new specific perspective in NAFLD therapeutic strategies.Evidence from meta-analysis indicates that a positive association between Conceptualization: Lin Wei, Hui-Guo Ding.Data curation: Lin Wei, Hui-Guo Ding.Formal analysis: Lin Wei, Hui-Guo Ding.Funding acquisition: Hui-Guo Ding.Methodology: Lin Wei.Project administration: Lin Wei, Hui-Guo Ding.Writing – original draft: Lin Wei.Writing – review & editing: Hui-Guo Ding.


Background: As society ages, the incidence of osteoporosis increases. In several studies, cadmium (Cd) is thought to be related to osteoporosis. However, there are conflicting reports about the relationship between Cd and the risk of osteoporosis and osteopenia. Therefore, the purpose of this meta-analysis was to explore the relationship between Cd and osteoporosis and osteopenia.Methods: Through a review of the literature, articles published in PubMed as of December 2020 were identified and the references of related publications and reviews were reviewed. Ultimately, 17 eligible articles were selected to determine the relationship between blood and urine Cd concentrations for the risk of osteoporosis or osteopenia. In this study, we performed a classification analysis, heterogeneity test, subgroup analysis, and evaluated publication bias.Results: A total of 17 studies were included, including seven on blood Cd and 10 on urine Cd. By combining the odds ratio (OR) and 95% confidence interval (CI) for the lowest and highest categories, the odds ratio of blood Cd concentration that increased the risk of osteoporosis or osteopenia was OR 1.21 (95% CI: 0.84–1.58) and that of urine Cd concentration that increased the risk of osteoporosis or osteopenia was OR 1.80 (95% CI: 1.42–2.18), and the results of the subgroup analysis were also consistent.Conclusions: Our research indicates that while urine cadmium (Cd) concentration may be related to increased risk of osteoporosis and osteopenia, blood Cd concentration may not. Therefore, compared to blood Cd concentration, urine Cd concentration may be more reliable as a risk factor for osteoporosis and osteopenia. This result should be interpreted with caution. Currently. research on the relationship between Cd concentration and osteoporosis and osteopenia is limited, thus, further large, high-quality prospective studies are required to elucidate the relationship between Cd concentration and osteoporosis and osteopenia. Osteoporosis is a systemic bone disease characterized by decreased bone mineral density, bone microstructure destruction, and increased risk of fragility fractures. Due to the high morbidity and mortality of diseases such as osteoporosis, fragility fractures, and other diseases, it has become a public health problem that needs to be solved urgently , 2. SomeCadmium (Cd) is a toxic non-essential transition metal. With the acceleration of global industrialization, Cd and its inorganic compounds are widely used in the manufacturing process of electroplating, batteries, pigments, plastics, and alloys. A large amount of Cd will enter the soil and, ultimately, the human body through contaminated food and water . Cd accuTherefore, the purpose of this meta-analysis was to explore the relationship between blood and urine Cd concentrations and the risk of osteoporosis or osteopenia.Ethical approval and written informed consent from patients were not necessary because our study was based on summaries and analyses of results of existing studies.Free keywords were used to search for articles published in PubMed till December 2020. The search words used were “cadmium,” “osteoporosis,” “osteopenia,” and “bone density.” In addition, in order to obtain further relevant literature, we manually searched the references for related articles.The articles were independently selected and commented on by two authors. First, the title and abstract were filtered based on the relevance of the topic. After reading the abstract, the full text was screened and articles that will eventually be included in the meta-analysis were selected. Articles that met the inclusion criteria were independently selected by two authors. When it was unclear whether an article should be included, a discussion was conducted with the third author to reach a consensus.The inclusion criteria were as follows: (1) studies including human subjects; (2) observational studies; (3) studies that reported the relationship between blood or urine Cd concentration and osteoporosis or bone mass loss, and (4) studies that calculated and reported relative risk (RR), odds ratio (OR), or hazard ratio (HR) and 95% confidence interval (CI) values.in vitro or laboratory studies; and (3) comments or case reports.The exclusion criteria were as follows: (1) Animal experiments; (2) Two examiners used standardized data collection forms to extract data independently. These differences were resolved through discussions with other investigators and referenced to the original article. The data extracted from each study included the first author's last name, publication year, study type, average age, male to female ratio, sample size, study country, bone density measurement method, blood or urine Cd concentration, adjusted variables, and the corresponding 95% CIs-OR estimate. If the OR value of different potential confounding factors was high, the OR value extracted reflected the maximum control of the potential confounding factors. When required, the authors of the preliminary study were contacted for more information.I2 statistics were used to assess the heterogeneity between studies . The comprehensive estimate of the risk of blood Cd concentration events did not change substantially without any research conducted through sensitivity analysis . The comprehensive estimate of the risk of urinary Cd concentration events did not change substantially after excluding any research conducted through sensitivity analysis  and Egger's test (P = 0.396) showed no publication bias, whereas for the relationship between urine Cd concentration and osteoporosis and osteopenia, the Begg's test  and Egger's test (P = 0.000) showed publication bias. In order to further evaluate publication bias, we adopted the method of pruning and filling. Through computer filling, the results after correction for publication bias did not change   .I2 = 59.9%, P = 0.058; female OR = 0.94, 95% CI: 0.44–1.44, I2=13.8%, P = 0.313). In addition, a subgroup analysis of the degree of osteoporosis was conducted, with the following categories: normal (T-score > −1.0), osteopenia (−2.5 ≤ T-score ≤ −1.0), and osteoporosis (T-score < −2.5). Blood Cd concentration was associated with an increased risk of osteoporosis  and osteopenia .The correlation between blood Cd concentration and osteoporosis and osteopenia was analyzed by subgroup analysis. The results are shown in I2 = 0.00%, P = 0.855; female OR = 1.62, 95% CI: 1.29–1.94, I2 = 25.1%, P = 0.22). For the analysis of the degree of osteoporosis, urine Cd concentration was associated with an increased risk of osteoporosis  and osteopenia .The correlation between urine Cd concentration and the risk of osteoporosis and osteopenia was analyzed using subgroup analysis. A subgroup analysis of sex revealed that urine Cd concentration was associated with an increased risk of osteoporosis and osteopenia for both males and females  concentration may be related to increased risk of osteoporosis and osteopenia, blood Cd concentration may not. Therefore, compared to blood Cd concentration, urine Cd concentration may be more reliable as a risk factor for osteoporosis and osteopenia. This result should be interpreted with caution. Currently. research on the relationship between Cd concentration and osteoporosis and osteopenia is limited, thus, further large, high-quality prospective studies are required to elucidate the relationship between Cd concentration and osteoporosis and osteopenia.The datasets presented in this study can be found in online repositories. The names of the repository/repositories and accession number(s) can be found in the article/supplementary material.BY and HL designed the meta-analysis. HL, MZ, and JM performed the literature retrieval and the data extraction. DL and LH contributed to writing the article. All authors read and approved the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


There exists a large inter-individual variation in the efficacy of valproic acid (VPA) against epilepsy. What are the genetic polymorphism influences of sodium channels on VPA response is still under discussion. In this study, a meta-analysis was used to further explore the effects of SCN1A and SCN2A gene polymorphism on VPA response in children with epilepsy.: The PubMed, EMBASE, Web of Science, Chinese National Knowledge Infrastructure, Chinese Science and Technique Journals Database, China Biology Medicine disc, and Wan Fang Database were searched up to April 2021 for appropriate studies regarding the association between SCN1A and SCN2A gene polymorphism on VPA response in children suffering from epilepsy. The meta-analysis was conducted by Review Manager 5.3 software.: The results of this meta-analysis will be submitted to a peer-reviewed journal for publication.: This meta-analysis will summarize the effects of SCN1A and SCN2A gene polymorphisms on VPA response in children with epilepsy.DOI 10.17605/OSF.IO/N2786. Antiepileptic drugs are commonly used for the treatment of epilepsy, but 30% of patients still relapse after obtaining adequate and adequate treatment. As a non-selective sodium channel blocker, sodium valproate can down-regulate the Scn3a gene expression of sodium channel through epigenetic pathway. Sodium valproate is a first-line drug for the treatment of epilepsy, and it has good effects on the control of generalized and local seizures, while these effects are greatly affected by genetic variation.,8Epilepsy is a common disease in the nervous system of children.,10 Mutations in SCN1A and SCN2A genes encoding the α subunit of ion channel are closely related to the response rate and drug resistance of valproic acid (VPA) and carbamazepine. Therefore, it is of great significance to improve the antiepileptic response rate of VPA from the point of genetics, the prognosis of patients and the rehabilitation of neurological functions.Voltage-gated sodium channel is a common target of antiepileptic drugs.–15 So far, no meta-analysis on the relationship between SCN1A and SCN2A polymorphisms and the response of VPA to childhood epilepsy can be detected. Therefore, we performed a meta-analysis to clarify the relationship between SCN1A and SCN2A polymorphisms and the efficacy of VPA in the treatment of childhood epilepsy.At present, the relationship between SCN1A and SCN2A polymorphisms and the efficacy of VPA in the treatment of epilepsy in children are still controversial.22.1The protocol of this review was registered in OSF (OSF registration number: DOI 10.17605/OSF.IO/N2786). According to some reports, it follows the statement guidelines of preferred reporting items for systematic reviews and meta-analyses protocol.2.2Literatures that meet the following criteria are included in this study:1.patients, aged less than 18 years old, were diagnosed as epilepsy by EEG and imaging examination.2.The study evaluated the effects of SCN1A and SCN2A polymorphism on the response to VPA treatment in children with epilepsy;3.cohort study;4.patients were treated with VPA;5.data on the relationship between polymorphism and the efficacy of VPA can be obtained or deduced from original articles or corresponding authors.2.3The exclusion criteria are as follows: case reports, meta-analysis, review articles, and studies without detailed genotype data.2.4Two researchers independently performed a systematic literature search in 7 electronic databases. The retrieval database includes the PubMed, EMBASE, Web of Science, Chinese National Knowledge Infrastructure, Chinese Science and Technique Journals Database, China Biology Medicine disc, and Wan Fang Database. The retrieval time was set to build the database until April 2021. The search strategy for PubMed is displayed in Table 2.52.5.1The flowchart is demonstrated in Figure 2.5.2In order to avoid bias in the process of data extraction, the 2 researchers adopted a blind method to extract data from the literature independently. The differences in the extracted data are figured out through negotiation and discussion. If necessary, some research data should be obtained directly from the corresponding authors. The following information was collected from each study: first author, year of publication, race, sample size, age, sex, efficacy evaluation criteria, genotypes, and the number of effective and ineffective people with different genotypes.2.5.3 The Newcastle-Ottawa scale values arrange from 0 to 9. Studies with the score of 6 are considered to be of high quality.We investigated the quality of each study based on the nine-point Newcastle-Ottawa Scale.2.5.4If the data are not complete in the article, only descriptive analysis should be conducted, rather than meta-analysis.2.5.5I2 statistics of Chi-Squared test. The significance level of heterogeneity threshold of Q test is 0.10, with P > .10. It can be considered that there is no heterogeneity among studies. I2 is used to measure the degree of heterogeneity among multiple research results. I2 values exhibit various degrees of heterogeneity in the range of 0% to 100% . As long as I2 is less than 50%, its heterogeneity is acceptable. When P < .10 or I2 > 50%, the random effect model of the DerSimonian&Laird method is adopted to calculate the combined statistics. Otherwise, the fixed effect model of the Mantel-Haenszel method is applied for calculation. When the P value was bilateral, with P < .05, it was considered to be statistically significant. All analyses were carried out using Review Manager 5.3 software.We calculated odds ratios and 95% confidence intervals for the analyses of the single-nucleotide polymorphisms and the efficacy of VPA. Five genotypic models, namely allele model (T vs C), heterozygote model (TC vs CC), homozygote model (TT vs CC), dominant model (TT + TC vs CC), and recessive model (TT vs TC + CC), were used to detect the relationship between single-nucleotide polymorphisms and response rate. The heterogeneity among studies was evaluated by performing Q-test and 2.5.6Through the review of the characteristics included in the study, the potential confounding factors were selected for subgroup analysis.2.5.7The eligible study was sequentially removed to perform the sensitivity analysis.2.5.8,20If no less than 10 studies are included, funnel charts are used to assess publication bias.2.5.9The content of this article does not involve moral approval or ethical review and would be presented in print or at relevant conferences.3,22 Voltage-gated sodium channel consists of one α subunit and several β subunits, in which α subunit is the main structural and functional unit. Mutations in SCN1A and SCN2A genes of α subunit are associated with epilepsy in children.,25 VPA may play an antiepileptic role by blocking sodium channels. Therefore, the mutation of sodium channel gene can lead to changes in the structure and function of the drug target, so that the drug cannot combine with the target, thus inhibiting the discharge of neurons and the emergence of drug resistance. Previous reports on the relationship between SCN1A and SCN2A gene mutations and the efficacy of VPA were inconsistent. In this study, meta-analysis was conducted to further explore the relationship between SCN1A and SCN2A gene polymorphisms and the efficacy of VPA in the treatment of childhood epilepsy, so as to provide etiological basis for individualized treatment of VPA in children with epilepsy.Voltage-gated sodium channel gene mutations may participate in the occurrence of epilepsy, thus affecting the response of individuals to antiepileptic drugs.Conceptualization: Lijiao Chen.Data curation: Zhuangfei Wen, Jiang Chen.Formal analysis: Zhuangfei Wen.Funding acquisition: Lijiao Chen.Methodology: Bin Zhu.Project administration: Lijiao Chen.Resources: Bin Zhu.Supervision: Lijiao Chen.Validation: Bin Zhu, Yan Lu.Visualization: Bin Zhu, Yan Lu.Writing – original draft: Lijiao Chen, Zhuangfei Wen, Jiang Chen.Writing – review & editing: Lijiao Chen, Zhuangfei Wen, Jiang Chen.


Meta‐analysis identified small but significant associations between IA and number of stool samples positive for all enteroviruses , consecutive positivity for enteroviruses  and number of stool samples positive specifically for enterovirus B . Virome analyses to date have demonstrated associations between enteroviruses and IA that may be clinically significant. However, larger prospective mNGS studies with more frequent sampling and follow‐up from pregnancy are required to further elucidate associations between early virus exposure and IA/T1D.Viruses are postulated as primary candidate triggers of islet autoimmunity (IA) and type 1 diabetes (T1D), based on considerable epidemiological and experimental evidence. Recent studies have investigated the association between all viruses (the ‘virome’) and IA/T1D using metagenomic next‐generation sequencing (mNGS). Current associations between the early life virome and the development of IA/T1D were analysed in a systematic review and meta‐analysis of human observational studies from Medline and EMBASE (published 2000–June 2020), without language restriction. Inclusion criteria were as follows: cohort and case–control studies examining the virome using mNGS in clinical specimens of children ≤18 years who developed IA/T1D. The National Health and Medical Research Council level of evidence scale and Newcastle–Ottawa scale were used for study appraisal. Meta‐analysis for exposure to specific viruses was performed using random‐effects models, and the strength of association was measured using odds ratios (ORs) and 95% confidence intervals (CIs). Eligible studies  included 1,425 participants  and examined IA ( Here, we report the first systematic review and meta‐analysis of observational studies using mNGS to investigate vertebrate‐infecting DNA and RNA viruses in children ≤18 years, and subsequent development of IA or T1D. Analysis of bacteriophage has been excluded from this review. The unbiased viral mNGS in early life and childhood has potential to comprehensively identify diabetogenic viruses increasing the IA/T1D risk or viruses affording protection. This may present new opportunities to intervene through antiviral medications or vaccination.22.1,This review is registered on PROSPERO (23 July 2020), registration number CRD42020188737. Two reviewers (Clare L. Faulkner and Yi Xuan Luo) independently conducted a systematic search for observational studies investigating the association between virome composition and/or abundance, and IA or T1D. EMBASE and MEDLINE databases were searched (2000–1 June 2020) using the strategy in Appendix Eligible studies were observational , using mNGS to characterise the virome in any clinical specimen in children aged ≤18 years who developed IA and/or T1D. Age restriction was imposed because IA often develops in childhood, suggesting viruses exert influence early in life. IA was defined as persistence of one or more autoantibodies against T1D‐associated autoantigens  in ≥2 time‐separated consecutive samples. Transplacental autoantibodies were excluded, defined as transient presence of the same autoantibody in a child <18 months and his/her mother. T1D was defined using American Diabetes Association criteria.n = 2), an independent advisor (Ki Wook Kim) was consulted to reach consensus decision. Case reports/series, uncontrolled studies, reviews and animal studies were omitted based on exclusion criteria.Two reviewers (Clare L. Faulkner and Yi Xuan Luo) screened titles and abstracts of identified studies Figure  and then2.2n = 4).Data extracted included publication authors, year and geographical location; study design; study participants; number of cases/controls; age; level of pre‐existing IA/T1D risk (human leukocyte antigen [HLA] genotype and family history); sample type, number and collection protocols; virus detection method and positivity threshold; rates of virus positivity in cases/controls; examined outcome (IA or T1D); measures of effect and funding. Original authors were contacted for insufficient or missing published data (Two reviewers (Clare L. Faulkner and Yi Xuan Luo) independently assessed the quality of included studies using the National Health and Medical Research Council (NHMRC) level of evidence scalep‐values for viruses present in children with IA or T1D versus controls from the extracted data using the Mantel–Haenszel method. Virus positivity was defined as virus material present in ≥1 study sample as detected by mNGS. Analysis was performed both for the number of case/control individuals and the number of case/control samples positive for virus. We used both fixed‐ and random‐effects models; only results from random‐effects models are presented due to heterogeneity of study populations. Statistical heterogeneity was explored using Cochrane's Q Test (χ2) and the I2 statistic, which indicate the proportion of variance of the summary effect attributable to between‐study heterogeneity. A p < 0.10 was considered a statistically significant heterogeneity, while I2 ≤ 25% and ≥75% were deemed low and high heterogeneity, respectively. Subgroup analyses were performed for geographical location, stool versus plasma, consecutive virus shedding and studies using comparable detection thresholds, and pooled ORs were calculated. Sensitivity and influence analyses were conducted by the study size. Data analysis was completed in Review Manager, Version 5.4 (Cochrane Collaboration),p ≤ 0.05.We calculated ORs with 95% CIs and  Q Test χ2 and theOur systematic review is reported using meta‐analysis of observational studies in epidemiology and Preferred Reporting Items for Systematic Reviews and Meta‐Analyses guidelines.3,,The search returned 778 publications (51 duplicates), leaving 727 articles for review. Title and abstract screening identified 12 publications for full‐text review. Four were excluded: three were repeat data sets3.1,Six studies defined IA as positivity for ≥1 T1D‐associated autoantibody; two defined as ≥2 autoantibodies . Stool versus plasma subgroup analysis gave pooled ORs 1.20  and 0.66 , respectively  and 0.87 , respectively.Seven studies investigated vertebrate‐infecting viruses and IA. No significant heterogeneity was observed, unless stated. Due to insufficient sample data, positivity for any virus was only analysed at the individual level, with no difference between cases and controls, pooled OR 1.03 . For stool versus plasma subgroup analysis, the pooled ORs were 1.14  and 0.80 , respectively . For stool versus plasma subgroup analysis, pooled ORs were 1.15  and 0.80 , respectively. There were minimal differences in effect sizes with sensitivity analysis.Meta‐analyses for specific viruses were conducted where proportions of positive case–control individuals or samples were reported in ≥1 study. Six studies found a significant association between the number of EV‐positive samples and IA . There were no associations between IA and EV‐A‐positive samples or individuals, pooled ORs 1.61  and 1.12 , respectively.Two studies,,p = 0.01; Figure p = 0.0007), but not EV‐A .Three stool studiesp = 0.20). However, for parechovirus‐positive samples and IA , there was significant heterogeneity . In influence analysis, the removal of the largest study outlier  strengthened the magnitude of association, pooled OR 0.44  and low heterogeneity . There was minimal difference in effect size for the number of individuals positive for parechovirus.Four studies reported parechovirus positivity in stool. There was no association between the number of individuals positive for parechovirus and IA, pooled OR 0.83 .Meta‐analyses at the individual and sample level for rotavirus, bocaparvovirus, anelloviruses, sapovirus, norovirus, cardiovirus, circovirus, mamastrovirus, mastadenovirus, kobuvirus, picobirnavirus, erythroparvovirus and roseolovirus showed no associations with IA Table . There w,Meta‐analysis of consecutive shedding of parechovirus, bocaparvovirus, anelloviruses and picobirnavirus found no associations with IA Table . Meta‐anStudy quality subgroup analysis was excluded as NOS scores were ≥8. HLA subgroup analysis was excluded as no studies stratified virus positivity by genotype, and all but twop = 0.02). Notably, human mastadenovirus F, astrovirus, human adenovirus 41, coxsackievirus A2 (CVA2), enteric cytopathic human orphan virus 30, coxsackievirus B3 and human parechovirus were more abundant in cases, while saffold virus, norovirus and rotavirus A were more abundant in controls. Every differentially abundant EV‐A  was more abundant in cases (p < 0.00001). Additionally, one studyp < 0.001).Only one studyp = 0.001).One analysis of very young children <6 months3.4,p = 0.83; and no heterogeneity χ2 = 0.21, p = 0.65, I2 = 0%).Two studiesp = 0.78) and human mastadenovirus F , with no heterogeneity between studies. One studyp = 0.05), but not individuals ; EV‐A was not associated with T1D. Limited studies precluded subgroup and sensitivity analyses.Meta‐analyses for EV, parechovirus, cardiovirus, norovirus, sapovirus, mastadenovirus, bocaparvovirus, mamastrovirus and anelloviruses were not significant for the number of positive individuals Table . Strain‐4This systematic review of 10 observational studies, involving 695 cases and 730 controls, demonstrated associations between virome composition in children ≤18 years and development of IA, but not T1D. This suggests early virome changes may influence initiation of IA, but not progression to T1D. There was a weak association between the number of stool samples positive for EV or EV‐B and IA, with approximately 1.2 times the odds of EV or EV‐B positivity in children who developed IA versus controls; ORs 1.14 (1.00–1.29) and 1.20 (1.01–1.42), respectively. There was 1.5 times the odds of consecutive EV shedding in stool of children with IA versus controls; OR 1.55 (1.09–2.20). Only one study measured consecutive shedding of EV serotypes,,,,,,,,These data suggest that specific gut vertebrate‐infecting viruses present in the gut virome, rather than the presence of any virus, influence IA risk. Research repeatedly reports associations between EV infection and IA initiation, supported by EV RNA in stools,,,We demonstrated an association between IA and the number of virus‐positive samples, but not individuals. Thus, when an individual has more than one positive sample, their IA risk is amplified. This may relate to viral persistence, protracted infection or increased exposure through reinfection. In particular, IA was associated with consecutive or prolonged EV/EV‐B shedding. Consecutive shedding is a strong indicator of persistent infection. Viral persistence is suggested to play a critical role in the development of autoimmunity,,,,The one study that conducted differential abundance analysis found 129 viruses with a ≥2‐fold difference in abundance in the gut of IA cases versus controls.4.1,To minimise bias, we implemented pre‐defined eligibility criteria, screening by independent reviewers, no language restrictions and sources beyond indexed databases. Random‐effects models may have provided more conservative effect estimates by accounting for study population heterogeneity and generating wider confidence intervals.,,Our findings have limitations. Only two studies examined T1D, limiting the analysis of associations between viruses and progression to T1D beyond IA initiation. Of the two studies that analysed T1D, one analysed prospectively collected samples and one analysed samples collected at/after T1D diagnosis. This may limit comparability due to potential differences when examining the virome after, rather than before, diagnosis of the study outcome. IA was predominantly defined as ≥1 autoantibody, despite single autoantibody conferring lower lifetime T1D risk versus multiple antibodies.,Studies applied different positivity thresholds, including 50p100K,,,Despite advances in metagenomics, virome analysis remains challenging, with rapid viral evolution complicating sequencingPositivity for viral nucleic acid is a marker of infection, not proof, as viruses may pass through the gut without productive infection, as with plant viruses and diet‐derived viruses.,,Studies sampled various body sites through stool, plasma and PBMCs. Direct comparison of across sites is difficult, requiring careful consideration of where viruses replicate. For example, EVs and mastadenoviruses preferentially infect and replicate at mucosal surfaces,Finally, this review did not examine bacteriophage to limit scope and adopt the precedent in other studies of focusing on only one class of virus to maximise detection sensitivity.χ2 and I2 limitations in detecting true heterogeneity. Studies were small: six had <50 participants, potentially causing small study effects. However, longitudinal sampling increased statistical power for detecting differences in the virome between cases and controls. Thus, we demonstrated the importance of comparing both number of virus‐positive individuals and samples. One studyOverall quality of included studies was high (NOS scores ≥8). All studies except one matched for ≥3 factors. Meta‐analysis demonstrated little significant heterogeneity; however, results must be interpreted cautiously given 4.2Despite limitations of targeted viral detection in IA/T1D pathogenesis studies, there remains a paucity of large, unbiased virome studies. Our findings must be validated in future studies that (1) include >200 participants and frequent longitudinal sampling preceding IA/T1D onset to improve statistical power and counter small‐study effects; (2) include a wider range of HLA genotypes to consider viral associations with IA/T1D in the context of genetic risk; (3) incorporate multicentre data to reduce geographical bias; (4) employ sensitive enrichment and comprehensive sequencing platforms; (5) integrate differential abundance analysis of viral load; and (6) sample various body sites to characterise viral strains and account for niche variations in viral abundance. Future studies should also include virome analysis during pregnancy to explore the role of antenatal and congenital infections in offspring IA/T1D development.This work was supported by the National Health and Medical Research Council Practitioner fellowship APP1136735 (to Maria E. Craig), the Juvenile Diabetes Research Foundation International Postdoctoral Fellowship 3‐PDF‐2020‐940‐A‐N and the Australian Diabetes Society Lindsey Baudinet Award (to Ki Wook Kim). The authors would like to thank Professors Ondrej Cinek and Kendra Vehik for providing supplementary raw data for inclusion in this systematic review.The author declares that there is no conflict of interests.Maria E. Craig and Ki Wook Kim designed the study and led the study group. Clare L. Faulkner performed the literature search, extracted, analysed and interpreted the data and wrote the manuscript. Yi Xuan Luo independently confirmed the database search and study quality. Ki Wook Kim acted as consultant in instances of uncertainty. Yi Xuan Luo, Sonia Isaacs, William D. Rawlinson, Maria E. Craig and Ki Wook Kim helped interpret the data. All authors were involved in the critical revision of this manuscript. Ki Wook Kim is the guarantor of this study.Supplementary Material 1Click here for additional data file.