An examination of the prospective associations with suicide attempt and risk was not possible due to a lack of studies. Several studies suggested that results could not be solely attributed to depressive symptoms. A high level of heterogeneity was observed in each meta-analysis. Moderation analysis was possible for gender and publication year only; neither was significant. Findings provide further evidence of a link between social anxiety and suicidal thoughts and behaviors in youth but are limited by the small number of studies of mixed quality. This review supports future research into social anxiety symptoms as potential risk factors and treatment targets for suicidal youth.Suicide is a leading cause of death among young people every year. Identifying risk factors provides opportunities to intervene, and social anxiety may represent one such factor. This systematic review and meta-analysis aimed to review the evidence of associations between social anxiety and suicidality in youth (10–25 years). Embase, PsycInfo, and Medline were searched to identify relevant articles. Meta-analysis was conducted to examine the mean effect sizes of concurrent and prospective associations between social anxiety and three indices of suicidality in adolescents aged 10–25 years. Meta-analyses of 16 studies showed that social anxiety was associated cross-sectionally with suicide attempt (The online version contains supplementary material available at 10.1007/s10802-022-00996-0. Almost 800,000 people die by suicide every year, and it is the fourth leading cause of death among young people globally  might be particularly relevant for suicide risk due to its overlap with key suicide risk factors. SAD is a common mental health condition, affecting about 11% of the population over the lifespan and typically starting in adolescence  prior to literature search. and (suicid* or self-injury or self-harm or self-mutil* or self-cut* or cutting or self-burn* or self-poison* or deliberate self-harm or DSH or parasuicid*) and (follow-up or follow up or longitud* or prospective or future or subsequent or epidemio*) and (child* or youth or adol* or young or teen*). We included the term parasuicide  to redut Welch, ), but stt Welch, .We followed the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines for the meta-analysis of observational studies. Studies were included if they (1) involved participants who were aged 10–25 years at the first assessment time point, (2) applied at least one measure for social anxiety diagnosis or symptoms and for a dimension of suicidality , (3) reported an outcome measure of the association between social anxiety diagnosis or symptoms and suicidality or examined factors that may underlie suicidality in people with social anxiety diagnosis or symptoms, and (4) were published in English language, peer-reviewed, and indexed scientific journals. Studies were excluded if (5) the study was a review article, a conference abstract or paper, or a research dissertation. Two researchers (EL and KC) independently examined all titles and abstracts to determine whether papers met inclusion criteria. Disagreement was resolved by consensus. Data was extracted independently using an electronic spreadsheet.Two researchers (EL and KC) assessed the risk of bias independently. The Effective Public Health Practice Project (EPHPP) Quality Assessment Tool for Quantitative Studies  was chosen as the effect size because it is commonly reported in observational studies. For studies that did not report r, standardized regression coefficients were converted to r . The Cochran’s Q test and the Higgin’s and Thompson’s I2 test were used to assess the degree of heterogeneity between studies. A statistically significant result from the Cochran’s Q test (p < 0.05) suggests the presence of heterogeneity. A higher I2 value indicates a higher degree of heterogeneity  suggests there is substantial asymmetry in the funnel plot, and such asymmetry may be indicative of publication bias. Moderator analyses were conducted to ascertain if sample characteristics impacted the effect size estimate. A series of meta-regressions was planned to examine several study characteristics as potential moderators when there were sufficient studies (k ≥ 5) in each subgroup: (1) age, (2) gender , (3) publication year, and (4) depressive symptoms.The risk of bias across studies was evaluated by inspecting the funnel plot and running the Egger’s test . Participants were between 12 and 24 years . Percentages of female participants ranged from 17 to 100% . Participants were recruited from a university (n = 3), a prison (n = 1), a hospital (n = 2), or the community via flyers, posters, or social media (n = 10). Three out of sixteen studies recruited clinical participants  = 38.74, p < 0.001, I2 = 84.5%. Neither gender nor publication year were found to be statistically significant moderators: gender (Q(1) = 0.01, p = 0.92), publication year (Q(1) = 0.01, p = 0.94). The moderator effects of depressive symptoms and age were not examined due to lack of studies: two studies reported the level of depressive symptoms in their study sample  = 207.74, p < 0.001, I2 = 98.6%.The mean effect size for the meta-analysis examining the association between T1 social anxiety and T1 current suicidal ideation was statistically significant, See Fig. . There wr = 0.24, p < 0.05, CI , suggesting that higher social anxiety is associated with higher current suicidal risk  = 49.35, p < 0.001, I2 = 95.9%.The mean effect size for the meta-analysis examining the association between T1 social anxiety and T1 current suicidal risk was statistically significant, Meta-analysis of T1 social anxiety with suicidal attempt in the follow-up period (T1 to T2) was not conducted as only one study examined this association.r = 0.62, p = 0.06, CI , suggesting that higher levels of social anxiety are not associated with higher T2 suicidal ideation  = 75.30, p < 0.001, I2 = 98.7%.The mean effect size for the meta-analysis examining the association between T1 social anxiety and T2 suicidal ideation was not statistically significant, See Fig. . There wMeta-analysis of T1 social anxiety with T2 current suicidal risk was not conducted due to the absence of relevant studies.ps > 0.05). As such, there is no evidence of publication bias.Visual inspection of the funnel plot suggests there was a lack of asymmetry in funnel plots for lifetime suicidal attempt See , Fig. 1.F = 3.99, p < 0.05). Another cross-sectional study reported a three-fold increase in risk of suicidal attempt in individuals with major depression who had a history of SAD . Using structural equation modelling, Gallagher et al. . Meta-analyses of 16 studies showed that social anxiety was associated cross-sectionally with suicidality, as measured by suicide attempt, suicidal ideation, and composite suicide risk, and prospectively at trend level with suicidal ideation. The examination of moderator effects was limited by a lack of studies. Only the effect of gender and publication year could be examined, with neither statistically significant, suggesting these factors cannot explain the heterogeneity between studies. Unfortunately the potential moderating effect of age could not be examined due to a lack of studies. It will be valuable to examine this in future studies, given the important potential developmental differences in the association between social anxiety and suicidality. There were several studies suggesting that results could not be purely explained by depressive symptoms but we were unable to test this quantitatively due to the limited number of available studies. Findings therefore provide evidence of a link between social anxiety and suicidal thoughts and behaviors in youth, but most of the studies are cross-sectional and only a few were available for each suicidality construct. This prevented us from drawing conclusions about causality, examining the unique contribution of depression, and explaining the stark heterogeneity among studies.The quality of the papers included was, on the whole, either weak or moderate, with only one study rated to be of good quality. The quality is partly compromised by the lack of statistical control for covariates, including other anxiety disorders. Thus, it remains unclear whether social anxiety specifically, or anxiety more generally, may increase the risk of suicide in this population. For example, it may be that individuals consider suicide because they feel unable to cope with the worry and anxiety they experience . Meta-analyses of 50 years of suicide research have indicated that the suicide risk factors evaluated in the literature has stayed somewhat consistent , with limited improvement in the ability to predict suicidal behavior over decades Below is the link to the electronic supplementary material.


The effect of HBV on neonatal and maternal outcomes can create a basis for more accurate clinical decision-making. So, the aim of this meta-analysis is to detrmine the effect of chronic hepatitis B virus on the risk of pregnancy outcomes by combining cohort studies.International databases in this meta-analysis included the Cumulated Index to Nursing and Allied Health Literature (CINAHL), SPORT Discuss via the EBSCO interface, PubMed (Medline), Scopus, Web of Science, Embase, which were searched up to April 2023. All cohort studies reporting the risk ratio (RR) with a 95% confidence interval (CI) were included in the study. The quality assessment was done based on the Newcastle–Ottawa Scale (NOS).P value: 0.00). Similarly, the association between the presence of hepatitis B infection in pregnant women and the occurrence of pre-eclampsia was 1.10 . The risk of preterm delivery in pregnant women with hepatitis B infection was 1.17 times that of pregnant women without hepatitis B infection .Finally, thirty-five cohort studies were selected for meta-analysis. Outcomes of interest included pre-eclampsia, gestational diabetes, abortion, preterm birth, infant death, and other related outcomes. Results showed that the pooled RR for incident gestational diabetes in pregnant women with choronic hepatitis B infection was 1.16 (RR: 1.16; 95% CI 1.13–1.18; I-square: 92.89%; This meta-analysis found that hepatitis B infection during pregnancy may be associated with an increased risk of gestational diabetes, preterm delivery, pre-eclampsia, and eclampsia. However, confirmation of this association, as well as the specific biological pathways involved in the association between HBV infection and pregnancy outcomes, requires further investigation.The online version contains supplementary material available at 10.1186/s12985-023-02182-0. A double-stranded DNA virus belonging to the hepadnaviridae family, hepatitis B virus (HBV). The virus is enveloped and has a core with a viral DNA genome of approximately 3200 base pairs. In patient blood, the virus was initially identified as “Australia antigen,” subsequently known as hepatitis B surface antigen (HBsAg). Later on, as a marker for those at high risk of transmitting the disease, the hepatitis B e antigen (HBeAg) was discovered. The virus penetrates a hepatocyte, delivers its genome to the nucleus, and changes the relaxed circular DNA into covalently closed circular DNA (cccDNA) –3. SigniThe effect of HBV on neonatal and maternal outcomes can create a basis for more accurate clinical decision-making. By identifying the specific connection between HBV and pregnancy outcomes, clinicians and specialists can reduce the impacts and enhance the quality of life for HBV patients. The research may aid in updating clinical guidelines and improving the care of HBV patients, which may aid in the early detection and prevention of pregnancy outcomes. Thus, by combining cohort data, we aimed to systematically review and meta-analyze the relationship between HBV and pregnancy outcomes.The current study was a systematic review and meta-analysis, which was conducted to determine the exact relationship between the presence of infection and the chronic HBV and the occurrence of maternal outcomes such as pre-eclampsia, gestational diabetes, abortion, premature birth, infant death, and other related results. The primary studies in this meta-analysis were prospective or retrospective cohort studies. All the steps of this study were developed and carried out based on the structure of the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guideline .At first, the search strategy aimed at retrieving cohort studies published in the target field in international databases such as the Cumulated Index to Nursing and Allied Health Literature (CINAHL), SPORT Discuss via the EBSCO interface, PubMed (Medline), Scopus, Web of Science, and Embase. Then the retrieved articles were screened in Endnote version 8. Keywords were included: “Hepatitis B”, “Hepatitis B Virus”, “Pregnancy Outcome”, “Maternal Morbidity”, “Maternal Death”, “Pre-Eclampsia”, “Premature Birth”, and “Gestational Diabetes”, along with their synonyms in the Mesh database. To perform the screening, first, the duplicates of retrieved articles in the software were removed, and then, in the first step, screening based on the title; in the second step, screening based on the subject; and in the final step, screening based on the full text of the articles was done. And after these steps, the final articles were selected. The time frame for searching international databases and screening was from January 1990 to February 2023. Also, all the screening steps were done independently by authors (MA and MA), and disputes were resolved by the third researcher (YM), who was an expert. To carry out a comprehensive and detailed search, or in other words, to carry out gray literature to complete the search strategy, the first ten pages of Google Scholar along with a manual search  were also done by the authors. The inclusion criteria and the final selection of studies in this meta-analysis were based on the PECOT structure, the specifics of which are detailed in Table The NOS  checklist was used to evaluate the quality of the articles. This checklist is designed to assess the quality of cross-sectional studies. Each of these items is given a score of 1 if they are observed in the studies. And the maximum score for each study is 9 points. This step was done independently by two authors (MA and KZ), and in case of disagreement, the cases were referred to the third researcher (YM).P-value was considered lower than 0.05. Subgroup analyses were performed to determine the main source of heterogeneity in the current meta-analysis based on gestational age, the continent or country of study, and mothers’ age.The intended effect size in this meta-analysis was the risk ratio (RR). First, the effect size and the confidence interval were extracted from the studies to perform the analysis. Then, in the desired software for analysis, the logarithm and the standard deviation (SD) of the RR logarithm were calculated, and by combining the logarithm and the standard deviation of the RR logarithm, meta-analysis was conducted. To check the heterogeneity and variance between the selected studies, Cochran’s Q and I2 tests were used. Statistical analysis was performed using STATA 17, and the In this meta-analysis, after searching and retrieving all articles, 1390 articles in the PubMed database, 1002 articles in the Scopus database, and 779 articles in other relevant databases, including the Cumulated Index to Nursing and Allied Health Literature (CINAHL), SPORT Discuss via the EBSCO interface, Web of Science, and Embase, were retrieved. After removing the duplicates that included 1850 articles, 1321 articles were screened based on the title. In this stage, 902 articles were removed based on the title, and 419 articles entered the screening stage based on the abstract and then the full text. Finally, a total of 384 articles were removed in these steps. Thirty-five cohort studies were selected for meta-analysis and the present study  .The first desired outcome in this meta-analysis was to estimate the incidence of GDM in pregnant women with HBV. The sample size was equal to 32,370,174 pregnant women in a total of 29 studies, of which 121,737 pregnant women were infected with HBV , 21–48. P value: 0.01) have a higher risk of developing HBV infection compared to pregnant women living in Asia   Table . Based o0) Table .P value: 0.00) .The second desired outcome in this meta-analysis was to estimate the risk of preeclampsia in pregnant women with HBV infection. The sample size was equal to 3,217,1538 pregnant women in a total of 20 studies, of which 103,392 were infected with hepatitis , 48–50. P value: 0.00). More than pregnant women living in Europe  and America . The important point is that HBV in pregnant Asian women has a positive and significant association with the occurrence of preeclampsia, while in European and American pregnant women, this relationship is inverse and protective. Based on maternal age and gestational age based on weeks, the results showed that an age below 30 years and a gestational age above 38 weeks aggravate the effect of HBV infection on the occurrence of preeclampsia, and the risk of preeclampsia is higher in these women  .The third desired outcome in this meta-analysis was the estimation of the risk of premature delivery in pregnant women with HBV infection. The sample size was equal to 34,950,154 pregnant women in a total of 27 studies, of which 132,441 pregnant women were infected with hepatitis , 50–54. P value: 0.00) is greater than that in pregnant women living in Europe  and Asia . Based on maternal age and gestational age by week, the results showed that gestational age lower than 38 weeks aggravates the effect of HBV infection on the occurrence of premature birth, and the risk of premature birth is higher in these women , but it was not statistically significant. If the risk of gestational hypertension , miscarriage : 78.96%; P value: 0.03), and neonatal death  in pregnant women with HBV infection was less than one. These results are reported in the Additional file Other outcomes examined in this meta-analysis included eclampsia, gestational hypertension, miscarriage, and neonatal death. The meta-analysis results showed that the risk of eclampsia in pregnant women with HBV infection was equal to 1.48 , which is observed in 5% of pregnant women. Preeclampsia has no specific symptoms and is dangerous for the fetus and the mother. The contraction of blood vessels causes this disease and, as a result, leads to an increase in blood pressure and a decrease in blood flow in fetal organs such as the liver, kidney, and brain. This reduction in blood flow in the uterus leads to problems for the fetus, such as reduced growth, reduced amniotic fluid, etc. The current meta-analysis showed that the presence of HBV infection could increase the risk of preeclampsia in pregnant women by 10% compared to pregnant women without HBV infection. Meta-analysis studies are conducted to determine the more accurate and error-free effect and relationship between two important factors. These studies can control many possible errors in the relationship between exposure and the desired outcome. The important point in studying the relationship between the presence of HBV infection and the occurrence of preeclampsia in pregnant women was the difference in the pathophysiology and epidemiology of preeclampsia and HBV infection in early studies conducted in the world. In addition, the statistical population and method of care for each of these conditions differed in these studies. Other infections, such as parasitic, viral, and bacterial infections, may have a significant impact on the incidence and prevalence of pre-eclampsia in pregnant women, and the presence of other blood-borne and sexually transmitted infections, particularly hepatitis C and HIV/AIDS, may also increase susceptibility to HBV infection. As a result, it is likely that the presence of these conditions and other infections predisposes women to pre-eclampsia, with HBV acting only as an aggravating or enabling factor , 55, 56.Subgroup analyses in determining the relationship between HBV infection and the occurrence of preeclampsia in pregnant women showed that the risk of infection is higher in Asian women. This result confirmed the difference in the effect of HBV on the occurrence of preeclampsia in different geographical regions, because in other regions such as Europe or America, the relationship between HBV infection and the occurrence of preeclampsia was an inverse or protective relationship. The main reason for this is the difference in culture, the way services related to prenatal care before, during, and after pregnancy are provided and received, and, most importantly, the prevalence of other infections that are effective in Asian countries or other locations. Finally, there are plausible reasons in clinical texts for the association between HBV and the development of pre-eclampsia. In general, HBV infection has been associated with an increased risk of atherosclerosis in pregnant women –60. FurtIn the present meta-analysis, results showed that the presence of HBV infection increases the risk of gestational diabetes in pregnant women by 16% compared to women without HBV infection. Also, the risk of developing gestational diabetes in Asian women with HBV infection was lower than that of American women. This difference can be attributed to the importance of the issue of pregnancy and the difference in receiving services related to gestational diabetes screening in the Asian region and Asian countries. In addition, the different diet  and the way of doing physical activity can be considered other reasons for this difference , 70. AnoThe important point of this meta-analysis was to estimate the effect size with high accuracy. Although the degree of heterogeneity in the estimated effect size was high, this degree of heterogeneity was indicative of statistical heterogeneity as determined by the I square index and Cochrane’s Q test. The important point was the absence of clinical heterogeneity or its presence at an acceptable level, which is interpreted by the estimated confidence intervals. All the confidence intervals obtained for the desired relationships in the present meta-analysis were narrow, which indicated high precision in the analysis. On the other hand, the narrow confidence interval is one of most important items in clinical interpretation and clinical justification of communication. For example, to determine the relationship between HBV infection and gestational diabetes, the estimated effect size was 1.16 with a confidence interval of 1.13 to 1.18. Despite the heterogeneity rate of 92.7%, the calculated confidence interval is very narrow, which indicates the existence of a sufficient sample and number of studies to determine the relationship and confirms the accuracy of the calculated relationship. Another strength of the present meta-analysis was the subgroup analysis based on the variables of geographic regions, gestational age, and the age of the pregnant mother, which to some extent, shows the role of confounding and other influencing variables in the relationship between HBV infection and pregnancy outcomes. The results of this meta-analysis can be very effective in developing or updating clinical guidelines.One of the weaknesses or limitations of the present meta-analysis is the failure to perform subgroup analyses based on important variables such as body mass index, the presence of other infections , other underlying diseases  pointed out that due to the lack of reporting of primary studies in their results, they were not included in the results of the present meta-analysis. We suggest that future studies be conducted to determine the relationship between HBV infection and the occurrence of pregnancy outcomes by considering these variables.This meta-analysis found that hepatitis B infection during pregnancy may be associated with an increased risk of gestational diabetes, preterm delivery, pre-eclampsia, and eclampsia. However, confirmation of this association, as well as the specific biological pathways involved in the association between HBV infection and pregnancy outcomes, requires further investigation. As a result, it is crucial to enhance programs and healthcare services for women in society, focusing on the promotion of screening, care, and treatment programs for infectious diseases, particularly HBV. These efforts should be implemented across various communities, with particular emphasis on developing societies.Additional file 1. Supplementary Figures.


Q and I2 tests were used to evaluate heterogeneity in the meta-analysis. Egger’s test was applied to assess publication bias. A total of 12 eligible studies were included in this meta-analysis; all were of moderate and high methodological quality. The pooled results suggested that increased TSH levels were significantly associated with PTC risk; however, the included studies were significantly heterogeneous. Stratification analysis indicated that the heterogeneity might be from the area or type of control. Although significant publication bias existed among the studies, the trim-and-fill method and sensitivity analysis revealed that the combined results were stable and robust. TSH levels are significantly associated with the PTC risk; however, more high-quality studies in large sample sizes are recommended to verify the extrapolation of these findings.High thyroid stimulating hormone (TSH) levels may stimulate papillary thyroid cancer (PTC) cell proliferation; however, the relationship between TSH levels and PTC risk remains controversial. We aim to ascertain the association through a meta-analysis. Literature searches were conducted in PubMed, Embase, and Web of Science databases. After literature screening, the methodological quality was assessed using the Newcastle-Ottawa Scale and Agency for Healthcare Research and Quality methods. Cochran’s   As thcurrence ,10. TherThyroid stimulating hormone (TSH) is a pituitary hormone whose serum or plasma levels are a sensitive marker for assessing hyper- or hypothyroidism . TSH is Therefore, this meta-analysis aimed to summarize observational studies on TSH levels and PTC risks and comprehensively assess and explore their association. This study’s findings will provide evidence for prediagnostic risk assessment of PTC to reduce invasive testing and overdiagnosis among patients.22.1Literature retrieval was carried out through PubMed, Embase, and Web of Science databases. Search keywords included “thyrotropin,” “thyrotropic hormone,” “thyroid stimulating hormone,” and “papillary thyroid cancer.” Keywords in the same category were joined together with “OR,” while keywords in different categories were joined together with “AND.” The search process combined subject terms with free terms, and the search strategy was adapted to the characteristics of the database. Detailed retrieval procedures of each database are displayed in Tables S1–S3. The search was available until January 12, 2023, without language restrictions. This study also screened relevant reviews and their included literature to expand the search scope of this meta-analysis.2.2Inclusion criteria: (1) the study reports TSH and PTC incidence risk, (2) the study type is cross-sectional, case–control, or cohort, and (3) the odds ratio (OR) and 95% confidence interval (CI) with univariate or multivariable adjustment are reported, or can be calculated by converting data based on frequency, sample size, etc.Exclusion criteria: (1) reviews, conference abstracts, and comments, (2) studies with participants under 18 years of age, and (3) for duplicate publications or multiple papers with the same batch of data, only the study with the most comprehensive data was kept, and the rest were excluded.2.3Two independent researchers conducted literature screening and data extraction based on the above criteria. Information to be extracted includes first author, publication year, region where the study was conducted, type of study, basic characteristics of the study population , TSH grouping, study outcomes, and correction factors. Any inconsistency in data extraction will be solved after discussion. The methodological quality of the case–control and cohort studies was appraised according to the Newcastle-Ottawa Scale (NOS) , which i2.4To assess the relationship between TSH and the risk of PTC, the OR and 95% CI were adopted as effect size indicators. If the included studies classified TSH into multiple categories, such as <0.40, 0.40–1.35, 1.36–2.12, 2.13–4.20, and >4.20 μIU/mL, then TSH <0.40 μIU/mL was used as the control group, and the OR (95% CI) of the remaining subgroups was combined in a random-effects model to calculate OR (95% CI) of ≥0.4 vs <0.40 μU/mL (high vs low) for the following meta-analysis. If the included studies used TSH as a continuous variable to calculate the risk of PTC occurrence, then this study used an increase of 1 μIU/mL as the dose change value for TSH and calculated its corresponding OR (95% CI) for the following meta-analysis.Q test and I2 test [P < 0.05 or I2 > 50%, and the random-effects model was therefore applied for the meta-analysis. Otherwise, the fixed effect model was adopted instead. Subgroup analysis was performed according to geographical location, study type, type of control, TSH grouping type, confounding factors adjusted or not, and methodological quality. A one-by-one elimination test [Cochran’s I2 test  were theion test  was usedion test  was usedion test  was empl33.1The detailed retrieval process and results are shown in 3.2As shown in The results of the quality assessment suggested AHRQ scores of 6–8 for cross-sectional studies and NOS scores of 5–9 for case–control studies (Tables S4 and S5). Thus, six of the 12 included articles are high-quality studies ,28–31,333.3I2 = 91.4% and P < 0.001, and the random-effects model was used in this meta-analysis. The pooled results estimated a significant association between elevated TSH levels and an increased risk of PTC, with OR (95% CI) = 1.70  and P < 0.001 , indicating that none of the grouping factors mentioned above contributed to the heterogeneity of combined results. In addition, except for the Western area and the control type of thyroid benign patients (TBP) or healthy control (HC), the results for all subgroups were consistent with the combined results, suggesting that elevated TSH levels may increase the risk of PTC .The subgroup analysis was conducted to identify sources of heterogeneity further . After s3.5P < 0.05, suggesting that the original pooled result was stable and not significantly altered by individual studies.Sensitivity analysis showed a change in OR (95% CI) from 1.50  to 1.83  for the combined results after eliminating articles one by one . MeanwhiP = 0.039 among the included studies. However, with the trim-and-fill method, the procedure did not fill in the dummy negative results to enhance the symmetry of the funnel plot, indicating that the meta-analysis results were stable.Furthermore, Egger’s test was conducted to ascertain publication bias, and the result suggested a significant publication bias at 4A recent study found that sensitivity to thyroid hormone indices was an indicator for predicting PTC . TSH is Although the above evidence suggests that the increased serum TSH level is correlated with cancer risk in thyroid nodules, the basis of this view is that TSH affects the proliferation of malignant transformed cells of differentiated thyroid cancers and thyroid cells . The strThe highlights of this study are as follows: (1) The effect values of this meta-analysis results were relatively large, suggesting a high strength of association between TSH levels and PTC risk. (2) The combined results do not alter significantly under the influence of any individual study, and the results are therefore robust. (3) Although the included studies exhibited significant publication bias, the trim-and-fill method suggested that the pooled results were stable. However, this study has some limitations: (1) The included studies were significantly heterogeneous, and subgroup analysis did not clarify the source of heterogeneity. (2) All included studies are observational studies with numerous confounding factors. Although most studies have conducted multi-factor correction, the corrected factors are not uniform, which may exaggerate the correlation strength between TSH and PTC. (3) The inconsistent grouping criteria of TSH levels may lead to clinical and statistical heterogeneity. It is hoped that follow-up studies can determine the uniform grouping criteria.5The combined results of this meta-analysis suggested a significant association between high levels of TSH and elevated risk of PTC. However, this relationship still needs to be supported by more evidence in the Western subgroup, as well as in the subgroup of controls containing healthy populations. Furthermore, more high-quality studies with larger sample sizes are recommended to validate the extrapolation of these findings.


This meta-analysis analytically reviewed recent studies concerning the potential associations between the brain-derived neurotrophic factor (BDNF) Val66Met polymorphism and susceptibility to major depressive disorder (MDD), with subgroup analyses for race and age.Relevant case-control studies were systematically searched for in PubMed, Embase, the Web of Science, China National Knowledge Infrastructure (CNKI), Wanfang, and Sinomed databases. A total of 24 studies were finally identified to have reported outcomes including alleles, dominant genes, recessive genes, homozygosity, and heterozygosity. Subgroup meta-analyses were performed based on participant age and ethnicity. Publication bias was represented by funnel plots. All meta-analyses of the randomized controlled trials included for evaluation were performed using RevMan5.3 software.P = 0.01). In the genetic model, dominant , recessive , and homozygous  genes were all associated with MDD.The findings revealed no significant association between BDNF Val66Met polymorphism and MDD. However, the Met allele was found to be associated with genetic susceptibility to MDD among white populations on subgroup analysis (OR = 1.25, 95% CI: 1.05–1.48, Despite the outcome limitations, this meta-analysis confirmed that the BDNF Val66Met polymorphism is a susceptibility factor for MDD in white populations. Major depressive disorder (MDD), also termed clinical depression, manifests with low mood, anhedonia, and fatigue; cognitive decline, autonomic nervous function changes, sleep disturbances, and appetite changes are also frequently reported , 2. MDD Recently, brain-derived neurotrophic factor (BDNF) gene polymorphisms have been reported to play important roles in depression and cognitive impairment , 7. As tThe BDNF gene is located on human chromosome 11p13. Of several known relevant gene polymorphisms, Val66Met has attracted the greatest interest. This non-synonymous G to A single-nucleotide polymorphism is found at position 196 of exon 2 and results in the substitution of methionine for valine  at codon 66 , 13. AltOn 10 September 2022, the Embase, PubMed, Web of Science, China National Knowledge Internet (CNKI), Wanfang, and Sinomed databases were searched for relevant literature using keywords as follows: (major depressive disorder OR depression disorder OR depression disease) and (brain-derived neurotrophic factor) and  and (polymorphism OR mutation OR variation OR genotype OR gene). Literature in both English and Chinese was searched.All studies included in our meta-analysis conformed to the following inclusion criteria: (a) the study focused on the relationship between MDD and Val66Met; (b) the study was a case-control study published in the form of a treatise; (c) the study included both MDD patients and normal controls; (d) study controls were healthy persons; (e) study sample sizes were adequate and allele and genotype frequency data were available; (f) the study applied the diagnostic criteria of DSM-IV or DSM-V or CCMD-3 or HDRS≥18 or ICD-10 MDD; and (g) the study employed rigorous experimental methods for detecting genetic polymorphisms. For duplicate publications, the study with the greater sample size was included in our analysis.All studies selected for analysis were sequentially screened on the basis of title and abstract content; studies which did not meet our selection criteria were excluded while duplicates were removed. The following data were extracted using pre-defined data extraction tables: first author's name; year of publication; country of study; study subject ethnicity; sources of both case and control groups; sample sizes; case and control group genotypes; allele frequencies; allele and genotype identification methods; whether Hardy-Weinberg equilibrium (HWE) conditions were met; and diagnostic criteria. Data evaluation was performed independently by Ou Li and Yuxia Wang; inconsistencies were addressed via research team discussions.2 test was used to evaluate heterogeneity among results of each study. When I2 ≤ 50%, studies were considered to have been homogenous and fixed-effects analysis was conducted. When I2 > 50%, studies were considered to have been heterogeneous and random-effects analysis was conducted. If the heterogeneity between study groups was too great, descriptive analysis was performed. Publication bias was analyzed using funnel plots. The impact of each study on pooled OR was assessed case-by-case.Data were analyzed using Revman 5.3 software. The odds ratio (OR) was used as a measure of BDNF gene polymorphism in MDD and a 95% confidence interval (CI) was considered. The χThe flow chart in P > 0.05 and I2 ≤ 50% for all genetic models, suggesting that there was little heterogeneity between the studies and all fixed-effects models were adopted for analysis. Analysis of Met allele data revealed a Val OR of 1.02 , confirming that there was no statistically significant difference between Met and Val alleles in MDD patients as compared to the healthy population. Meta-analysis findings based on different genetic models revealed dominant, recessive, heterozygous, and homozygous gene model ORs to be 1.08 , 1.05 , 0.98 , and 1.01 , respectively, confirming that differences between MDD patients and healthy controls were not statistically significant for each gene model. Thus, no significant association of the Val66Met polymorphism with MDD was determined  for the Caucasian allele Met:Val, suggesting that Val allele frequency was reduced while Met allele frequency was increased in white MDD patients as compared to controls. Thus, the Met allele was found to be associated with a genetic susceptibility to MDD among white populations. Meta-analysis based on different genetic models revealed dominant, recessive, heterozygous, and homozygous gene model ORs for white populations to be 1.40 , 1.70 , 1.20 , and 1.77 , respectively. Statistically significant differences between dominant, recessive, and homozygous gene models were found. These three gene models were found to be associated with a genetic susceptibility to MDD in Caucasian populations. The results of the Asian populations subgroup meta-analysis were not statistically significant and consequently the Val66Met polymorphism was not determined to associate with a genetic susceptibility to MDD in Asian populations  has defined age categories for people, among which elderly people are considered to be those over 60 years old. Age-based subgroup analysis was performed considering 60 years of age as a cut-off for dividing the subjects into two groups. Heterogeneity testing revealed that y to MDD .Sensitivity analysis revealed that risk estimates were not significantly affected upon exclusion of any individual study, confirming that the results of this meta-analysis were reliable. According to genotype indicator, a funnel plot analysis of potential publication bias revealed a roughly symmetrical distribution of values and no significant pattern asymmetry, suggesting that the publication bias in the literature we analyzed was negligible .The pathophysiological mechanisms via which the BDNF Val66Met gene polymorphism acts to induce MDD remain unclear. A dimeric, neurotrophic factor primarily concentrated in the hippocampus and frontal cortex, BDNF is closely related to neuronal plasticity, differentiation, survival, and repair . ImportaThe literature used in our study on Caucasian populations included 11 articles, five from the United States and remaining six from Poland, Spain, Turkey, Germany, Israel, and Australia. The literature on Asian populations included 13 articles, 11 of which were from China and one each from the Republic of Korea and India. In this work, we retrieved 24 studies including 3,497 cases and 5,623 controls, thus studying a relatively large sample size that certainly reduced the probability of false negatives. Our findings revealed that the total effect of the BDNF Val66Met gene polymorphism was not significantly associated with MDD. However, a subgroup analysis revealed that the Met allele was associated with a genetic susceptibility to MDD in Caucasian populations, with statistically significant differences between dominant, recessive, and homozygous genotypes. Age was not found to associate with any susceptibility to MDD. No publication bias was noted in our study.The present study suggests BDNF Val66Met gene polymorphism plays a role in major depressive disorder in Caucasian populations but no such pattern was found in Asian populations. We considered that the possible reasons for the conflicting results include changes in the frequency of the same gene mutation between populations, the impact of the same gene mutation on different races and ages, and the impact of different environments. The frequency of Met alleles in white populations is 25–32% , while tThis meta-analysis has some limitations. Since our study only included literature published in Chinese and English, the possibility of publication bias cannot be ruled out. As the number of available case-control studies was relatively small, future meta-analyses evaluating larger samples are required to draw more accurate conclusions. Due to limited data, we were unable to stratify analysis according to factors such as years of education, sex, and exposure to various environmental factors. More precise assessments of the literature should be conducted in future meta-analyses.The original contributions presented in the study are included in the article/supplementary material, further inquiries can be directed to the corresponding author.NL, ZZ, and ZS were responsible for the literature collection and collation. YW and OL processed the data and wrote the paper. JX revised the article. All authors contributed to the article and approved the submitted version.


The relationship between red and processed meat consumption and pancreatic cancer risk is controversial and no study has looked specifically at the correlation for 6 years. We conducted a meta-analysis to summarize the evidence about the association between them.We systematically searched PubMed, Embase and Cochrane Library for studies of red or processed meat consumption and pancreatic cancer published from December 2016 to July 2022. We performed random-effects models to pool the relative risks from individual studies. Subgroup analyses were used to figure out heterogeneity. We also performed publication bias analysis.p = 0.064) for red meat and 1.04  for processed meat with statistically significant heterogeneity.Seven cohort studies and one case–control study that contained a total of 7,158 pancreatic cancer cases from 805,177 participants were eligible for inclusion. The combined RRs (95% CI) comparing highest and lowest categories were 1.07 (95% CI: 0.91–1.26; This meta-analysis suggested that red and processed meat consumption has no relationship with pancreatic cancer risk. Pancreatic cancer, ranking as the seventh leading cause of mortality from malignancies globally owing to its unfavorable prognosis, exhibits a higher incidence rate in nations with a High Human Development Index (HDI) countries . This trStrong evidence suggests that consumption of red and processed meat may elevate the risk of pancreatic cancer due to its high content of heterocyclic amines (HCAs), polycyclic aromatic hydrocarbons (PHAs), and N-nitroso compounds (NOCs). In April 2017, Zhao et al. conducted a systematic review and meta-analysis investigating the association between intake of red and processed meat and the risk of pancreatic cancer. The analysis included a total of 28 cohort and case–control studies published before February 2016. Among the case–control studies, higher consumption of red meat and processed meat exhibited a positive correlation with pancreatic cancer risk ; however, no overall association was observed in cohort studies . Dose–response analysis indicated that every additional daily intake increment by 100 g in red meat was associated with an 11% increased risk for developing pancreatic cancer while each 50 g/d increase in processed meat intake led to an 8% rise in this risk. The findings from this study align with dietary guidelines while providing more comprehensive evidence based on robust epidemiological data. However, no clear relationship has been established between recent six-year trends in red or processed meat consumption and pancreatic cancer risk.Therefore, we performed an updated meta-analysis incorporating epidemiological studies published from February 2016 to July 2022 (including cohort and case–control studies) to investigate associations between red or processed meat consumption and incidence or mortality of pancreatic cancer. Additionally, the potential influence of factors such as sex (men vs. women), geographic area , duration of follow-up (less than 20 years vs. more than 20 years) and adjustments for alcohol intake, smoking, BMI, diabetes, family history pancreatic cancer, energy intake and physical activity were investigated.Articles published from December 2016 to October 2022 were systematically searched in PubMed, Embase and Cochrane Library. The search strategy was a combination of medical subject headings and free text words. Following medical subject terms were used: “Pancreatic Neoplasms,” “Red Meat” and “Meat Products.” Free text words including “red meat*,” “beef,” “pork,” “lamb,” “mutton,” “veal,” “pancreatic neoplasm*,” “Cancer of Pancreas,” “pancreas cancer*,” “pancreatic cancer*,” “Cancer of the Pancreas,” “diet*,” “food*” and “processed meat.” Moreover, we reviewed the reference lists from the included articles and those from previous meta-analysis to identify additional relevant studies.Studies were included in our meta-analysis when they meet the following criteria:1. Be a cohort or case–control study.2. Provide the 95% confidence intervals and adjusted estimates of the relative risk (RR)  for red and/or processed meat of pancreatic cancer incidence or mortality.3. The study was published from December 2016 to July 2022.4. The study was published in English. We considered “ham,” “sausage,” “bacon,” “salami” and “hot dogs” as equivalent to “processed meat.” When there were multiple published reports from the same study population, we chose the one with the largest population. For studies that researched for more than 1 cohort, the data of each cohort were selected  (4).Two investigators independently extracted the following data from each study: the first author’s name, the year of publication, the country where the study was performed, age of subjects, number of participants, follow-up time, type of the meat and their consumption strategy, adjusted RR/OR/HR with the corresponding 95% CI (highest to lowest) and other influencing factors. When some studies gave the RR values of processed red meat and unprocessed red meat, we extracted unprocessed red meat as the data of red meat in that study. When a study provided RR values for all age groups vs. people aged 50 years, we extracted data for all age groups  (5).The Newcastle-Ottawa Scale (NOS) was applied to assess the study quality , which rDue to variations in the reporting methods of effect size and 95% CI across studies, we estimated these values by comparing the highest with the lowest consumption categories . For simplicity, we use the term RR (relative risk) for all estimates in this manuscript since hazard ratios in some cohort studies and odds ratios in case–control studies are closely approximate relative risks , 8 when 2 statistics were used to evaluate statistical heterogeneity among studies , geographic area , duration of follow-up (<20 years vs. >20 years) and adjustments for alcohol intake, smoking, BMI, diabetes, family history pancreatic cancer, energy intake and physical activity.Q and I studies , when p To evaluate the stability of the results, we performed sensitivity analyses of statistically significant results, which were more stable if the pooled effect size changed little and none of the 95% confidence intervals crossed 1 after removing a single study.p < 0.05 considered statistically significant.We also used Egger’s regression asymmetry test  and BeggAll statistical analyses were performed with STATA version 17.0 software .The literature search generated 471 records, of which 109 were considered potentially valuable . FinallyThe characteristics of included studies are outlined in I2 = 49.5%, p = 0.064)  with statistically significant heterogeneity observed (= 0.064) .Subgroup analyses are shown in p = 0.438) and Begg’s test (p = 0.230) did not detect publication bias   with statistically significant heterogeneity observed (= 0.006) .Subgroup analyses are shown in p = 0.932) and Begg’s test (p = 0.806) did not detect publication bias . Af. Af16). Our findings are consistent with those of numerous other meta-analyses that have reached similar conclusions regarding the association between red and processed meat consumption and pancreatic cancer risk. For instance, Farvid et al. performed a systematic review and meta-analysis of 148 prospective studies published up to December 2020 to summarize the evidence on the relationship between consumption of red meat (unprocessed), processed meat, total red meat, processed meat, and the incidence of various cancers. By utilizing random-effects models to summarize relative risks for the highest and lowest intake categories for each exposure variable, they concluded that neither intake of red meat alone (unprocessed) nor processed meat alone was associated with an increased risk of pancreatic cancer . Han et However, most studies suggest that there is a biologically plausible positive association between the consumption of red and processed meat and pancreatic cancer. Several mechanisms have been proposed to explain the potential role played by red and processed meat in increasing pancreatic cancer risk. For instance, it is well-established that red meat contains heme iron, which stimulates the production of N-nitroso compounds (NOCs) by bacteria in the large intestine . MoreoveOur study possesses several strengths. Firstly, our search strategy was meticulously detailed. Secondly, it was conducted by two independent reviewers for data selection and extraction, with consultation from a third investigator in case of disagreements after discussion, thereby minimizing bias and error. Thirdly, we have a substantial sample size that allows for more robust conclusions regarding the association between intake of red and processed meat and pancreatic cancer risk. Fourthly, the studies included in this article were predominantly cohort studies (with only one being a case–control study), which reduces the potential for recall and selection biases. Lastly, subgroup analyses were performed to investigate the sources of heterogeneity.However, our study also has certain limitations. Firstly, the original studies encompassed both cohort studies and case–control studies; therefore recall bias, interviewer bias, and inaccurate measures of dietary consumption in case–control studies may impact the outcomes concerning processed meat consumption and pancreatic cancer risk. Secondly, there might be selection bias within the study population as it primarily focused on two high human development index (HDI) countries, the United States and Italy, where a higher incidence of pancreatic cancer is associated with a higher quality of life. Additionally, individuals in these countries tend to consume more red and processed meat compared to those in developing nations which could potentially inflate associations observed. Moreover, some participants were recruited from health care registries who generally exhibit greater attention toward healthy living practices thus reducing correlations between variables studied here. Thirdly, due to variations among original studies regarding specific daily meat consumption details, the inclusion criteria uniformly extracted highest and lowest levels of relative risk (RR) intake. Fourthly, certain studies accounted for potential confounding factors, including gender, alcohol consumption, smoking habits, and body mass index (BMI), while others did not. Fifthly, it is worth noting that our findings may have been influenced by imprecise measurements of meat intake in the original studies or variations in meat cooking methods, which could have impacted the overall relative risk estimation. Furthermore, since the original studies included in our study employed different unit categories , standardization was not feasible for extracting average values for red meat or processed meat from each stud; thus preventing us from conducting a dose–response analysis. We hope that future studies can provide more detailed investigations into the dose–response relationship between red/processed meat consumption and pancreatic cancer.Our meta-analysis of cohort and case–control studies revealed no significant association between red meat and processed meat consumption and pancreatic cancer risk. However, considering the dietary guidelines proposed by the NutriRECS consortium in 2019  and concThe original contributions presented in the study are included in the article/YaS and YuS designed the study. YuS contributed to the literature database search, data collection, data extraction, data analysis, and writing of the manuscript. XH performed the data extraction and check of the results. All authors contributed to the article and approved the submitted version.


Recent studies have suggested a relationship between gut microbiota and non-alcoholic fatty liver disease (NAFLD)/nonalcoholic steatohepatitis (NASH). However, the nature and direction of this potential causal relationship are still unclear. This study used two-sample Mendelian randomization (MR) to clarify the potential causal links.Summary-level Genome-Wide Association Studies (GWAS) statistical data for gut microbiota and NAFLD/NASH were obtained from MiBioGen and FinnGen respectively. The MR analyses were performed mainly using the inverse-variance weighted (IVW) method, with sensitivity analyses conducted to verify the robustness. Additionally, reverse MR analyses were performed to examine any potential reverse causal associations.P =0.04), Enterobacteriaceae (P =0.04), Lachnospiraceae UCG-004 (P =0.02), and Prevotella9 (P =0.04) and increased risk of NAFLD. Dorea (P =0.03) and Veillonella (P =0.04) could increase the risks of NASH while Oscillospira (P =0.04) and Ruminococcaceae UCG-013 (P=0.005) could decrease them. We also identified that NAFLD was found to potentially cause an increased abundance in Holdemania (P =0.007) and Ruminococcus2 (P =0.002). However, we found no evidence of reverse causation in the microbial taxa associations with NASH.Our analysis, primarily based on the IVW method, strongly supports the existence of causal relationships between four microbial taxa and NAFLD, and four taxa with NASH. Specifically, associations were observed between Enterobacteriales (This study identified several specific gut microbiota that are causally related to NAFLD and NASH. Observations herein may provide promising theoretical groundwork for potential prevention and treatment strategies for NAFLD and its progression to NASH in future. NonalcoNAFLD/NASH results from an interplay of multiple factors, including a genetic predisposition, obesity, insulin resistance, and an inflammatory cascade reaction . In the Currently, research indicates that the composition and abundance of gut microbiota change significantly at the phylum, class, family, and genus levels among NAFLD/NASH patients, compared to healthy individuals . In addiMendelian randomization (MR), which is a tool leveraging genetic variants linked to modifiable exposures to investigate their potential causal relationship with outcomes, offers a possible solution to the issues of confounding and reverse causality. It has already been applied in studying the relationship between gut microbiota and various diseases . Our pri22.12.1.1The human gut microbiome genome-wide association study (GWAS) dataset utilized in this research was sourced from the largest meta-analysis of microbiota composition to date, conducted by the MiBioGen consortium . The MiB2.1.2The GWAS summary statistics used in this research for NAFLD and NASH were obtained from the FinnGen Consortium R9 release data, which were made publicly available in May of 2023. The NAFLD GWAS summary statistics were derived from a cohort of 377,277 Finnish adults, comprising 2,275 cases and 375,002 controls. GWAS summary statistics for NASH included 157 cases and 377,120 controls. Further information regarding the cohorts, genotypes, endpoint definitions, and association tests used in the FinnGen consortium is available through the FinnGen webpage.2.2P <10-5, as reported by Sanna et al. (2019) . Additionally, we excluded SNPs with minor allele frequencies (MAF) <0.01 and palindromic SNPs. Finally, we employed the F-statistic to evaluate the strength of the instrumental variables (IVs). This metric is derived using a formula [F = R2×(N-k-1)/k×(1-R2)] that accounts for a variety of factors, including total variance (R2), sample size (N), and the number of IVs included in the analysis (k)  . Secondlysis (k) . Weak geysis (k) .2.3Mendelian randomization analysis, a cutting-edge methodology that has been gaining traction in recent years, owing to its unique ability to circumvent the limitations of traditional research methods. At its core, MR analysis leverages exposure-related genetic variation as IVs, in order to investigate the causal effects of modifiable exposures on outcomes . Thanks To ensure the validity of our results, we had to adhere to a rigorous set of assumptions shown in Figure 1In this study, we employed five different MR methods, including inverse-variance weighted (IVW), weighted median (WM), MR‐Egger, simple mode, and weighted mode, to assess the causal effects of gut microbiota on NAFLD and NASH. These MR methods have varying assumptions regarding horizontal pleiotropy. Various statistical methods were used in this study. As the IVW method is used to estimate causality through a meta-analysis of Wald values, the IVW method was used as the main method . In ordeP <0.05 was employed, with MR estimates presented as odds ratios (OR) with 95% confidence intervals (CI). Overall, the five different MR methods utilized in this study were chosen based on their underlying assumptions regarding horizontal pleiotropy, and were employed to investigate the causal relationships between gut microbiota and disease.The WM method was used to exclude invalid instruments, with a maximum allowable proportion of 50% . In conthttp://www.phenoscanner.medschl.cam.ac.uk/). To avoid potential confounding factors from impacting the ultimate causality, we eliminated SNPs linked to NAFLD or NASH risk factors and subsequently re-conducted the MR analysis.Furthermore, we conducted an evaluation of whether IVs were correlated with risk factors for NAFLD and NASH using phenoscanner  .33.1The baseline information for NAFLD, NASH, and microbiome cohorts was shown in 3.23.2.1Enterobacteriaceae, Lachnospiraceae UCG-004, Prevotella7, Prevotella9, Desulfovibrionales, and Enterobacteriales, exhibited significant associations with increased or decreased risk of NAFLD in terms of relative abundance. Initially, our preliminary MR analysis revealed an inverse relation between the risk of NAFLD and the relative abundance of Gammaproteobacteria as predicted genetically . This protective causality became insignificant after we removed a single SNP (rs75101789) using phenoscanner . Genetically predicted Enterobacteriaceae abundance was positively linked to an increased risk of NAFLD , with this causal relationship remaining significant even after excluding SNPs associated with NAFLD risk factors . Higher gene-predicted Lachnospiraceae UCG-004 abundance was also associated with elevated NAFLD risk , with consistent findings observed in the repeated MR analysis that excluded confounding SNPs . The original MR analysis suggested that people with higher abundance of Prevotella7 have lower NAFLD risks , but this causal relationship became less significant in the repeated MR analysis after excluding confounding SNPs . However, higher gene-predicted Prevotella9 abundance was significantly associated with increased NAFLD risk , with this causal relationship remaining significant even after excluding SNPs linked to NAFLD risk factors . The increased abundance of Desulfovibrionales may reduce the risk of NAFLD  following MR analysis, with consistent findings observed in both IVW and WM tests. However, this protective causality did not hold after excluding confounding SNPs . Finally, genetically predicted Enterobacteriales abundance was associated with an elevated risk of NAFLD , with this causal relationship remaining significant even after excluding SNPs associated with NAFLD risk factors . The gut microbiota obtained after screening for causal association with NAFLD are shown in The results of the IVW test indicated that the genetically predicted abundance of seven microbial taxa, specifically Gammaproteobacteria, Holdemania, genus Ruminococcus2). The reverse MR analysis showed that NAFLD could increase the abundance of the genus Holdemania  and genus Ruminococcus2 , and these results were shown in In the reverse MR analysis, no evidence of a causal effect of NAFLD on the aforementioned microbial taxa was observed. This suggests that the results of our MR analyses are not affected by reverse causation. Interestingly, in the reverse MR analysis, we found that NAFLD appeared to influence the abundance of 2 gut microbial taxa , Veillonella, Rhodospirillales) and NASH. Specifically, increased relative abundance of Dorea was found to be associated with an increased risk of NASH, as determined by the primary outcome of the IVW method , with similar results obtained from other methods such as WM. This causality remained significant even after excluding SNPs associated with risk factors . Conversely, raw MR analysis suggested that the increased abundance of Oscillospira may reduce the risk of NASH , a relationship that remained significant after excluding SNPs associated with risk factors . Our study also showed that the increased abundance of Ruminococcaceae UCG-013  and Ruminococcus1  can reduce the risk of NASH, whereas Ruminococcaceae UCG-014  was a risk factor for NASH. However, the genetically predicted Ruminococcaceae UCG014  and Ruminococcus1  showed an insignificant relationship with NASH after excluding confounding factors while Ruminococcaceae UCG-013 still showed a significant relationship with NASH . Additionally, an unknown genus (id.2041) appeared to be protective against NASH, but subsequent MR analyses did not yield significant results . Finally, we found that elevated relative abundance of Veillonella  and Rhodospirillales  was associated with an increased risk of NASH. However, after excluding confounders, this causal relationship between Veillonella and NASH remained significant , but this causal relationship betewen Rhodospirillales and NASH was not statistically significant . The gut microbiota obtained after screening for causal association with NASH are shown in The preliminary MR analyses revealed a causal relationship between the relative abundance of eight genetically predicted gut microbial taxa  and NAFLD/NASH. Order Enterobacteriales, family Prevotella is a genus of Gram-negative bacteria that is part of the Bacteroidetes phylum and it is one of the most abundant microbiota in the human gut   compared to those with F0/F1 fibrosis  and the Homeostatic Model Assessment for Insulin Resistance (HOME-IR) levels  . Based olmonella , has sublmonella . Shen etfibrosis . The gen) levels , while b) levels . Meanwhitrations .Dorea, Oscillospira, Ruminococcaceae UCG-013, and Veillonella and NASH through MR analysis.NASH, a potentially severe stage of NAFLD that may progress to cirrhosis and hepatocellular carcinoma , is tighOscillospira decreased and proportion of Dorea increased in NASH group through multivariate analysis  was extremely small, a relatively lenient threshold (P <1×10−5) was adopted for screening IVs. (ii) This study included individuals of essentially European ancestry, so extrapolating the findings to other populations is limited. (iii) Summary statistics for gut microbial taxa was profiled by targeting three distinct variable regions of the 16S rRNA gene. Bacterial taxa were only analyzed at the genus level but not at a more specialized level such as species or strain levels. When microbiota genome-wide association studies use more advanced shotgun metagenomic sequencing analysis, the results can be more specific and accurate. (iv) The GM-related GWAS summary-level dataset included in this study was based on 16S rRNA sequencing. Therefore, further analysis based on large-scale studies with more advanced methods, such as metagenomics sequencing, is required in the future to evaluate at the species-level. (v) Since the MR analysis is based on an untestable assumption, further experimental and clinical validation studies are crucial to test the clinical significance of microbial species.It must be acknowledged that there are some deficiencies in our research. (i) Since the number of IVs fulfilling the strict threshold (In summary, we performed a bidirectional two-sample MR analysis using published GWAS summary data to identify 6 types of microbial taxa that contribute to the development of NAFLD/NASH and also identified 2 types of microbial taxa that causally decrease NASH risk. In addition, we determined that NAFLD can affect the abundance of gut microbiota reversely. Future studies are warranted to further dissect the potential mechanisms of action between specific microbial taxa and NAFLD and NASH.https://www.finngen.fi/en/access_results and https://mibiogen.gcc.rug.nl.Publicly available datasets were analyzed in this study. This data can be found here: Ethical approval was not required for the studies involving humans because the original dataset used is fully public and allowed to be used, and all have been ethically approved. The studies were conducted in accordance with the local legislation and institutional requirements. Written informed consent for participation was not required from the participants or the participants’ legal guardians/next of kin in accordance with the national legislation and institutional requirements because the original dataset used is fully public and allowed to be used, and all have informed consent.HW: Conceptualization, Data curation, Investigation, Visualization, Writing – original draft. QZ: Data curation, Methodology, Software, Visualization, Writing – original draft. SZ: Data curation, Investigation, Validation, Writing – review & editing. TZ: Data curation, Writing – original draft. CD: Methodology, Writing – original draft. JY: Investigation, Writing – original draft. JP: Investigation, Writing – original draft. CC: Investigation, Project administration, Writing – review & editing. JL: Funding acquisition, Project administration, Validation, Writing – review & editing.


To investigate the association between circulating lipoprotein(a) (Lp(a)) and risk of all-cause and cause-specific mortality in the general population and in patients with chronic diseases, and to elucidate the dose-response relations.2: 75.34%, n = 19) in the general population and 1.18  in patients with cardiovascular diseases (CVD). The HRs for CVD mortality were 1.33  in the general population, 1.25  in patients with CVD and 2.53  in patients with diabetes mellitus. Linear dose-response analyses revealed that each 50 mg/dL increase in Lp(a) levels was associated with 31% and 15% greater risk of CVD death in the general population and in patients with CVD. No non-linear dose-response association was observed between Lp(a) levels and risk of all-cause or CVD mortality in the general population or in patients with CVD (Pnonlinearity > 0.05).We searched literature to find prospective studies reporting adjusted risk estimates on the association of Lp(a) and mortality outcomes. Forty-three publications, reporting on 75 studies , were included. The hazard ratios (HRs) and 95% confidence intervals (95%CI ) for the top versus bottom tertile of Lp(a) levels and risk of all-cause mortality were 1.09  levels are associated with higher risk of all-cause mortality and CVD-death in the general population and in patients with CVD. These findings support the ESC/EAS Guidelines that recommend Lp(a) should be measured at least once in each adult person’s lifetime, since our study suggests those with higher Lp(a) might also have higher risk of mortality.The online version contains supplementary material available at 10.1007/s10654-022-00956-4. Lipoprotein(a) (Lp(a)) consists of a cholesteryl-ester rich lipid core and an apolipoprotein B-100 bound to apolipoprotein(a) (apo(a)). Lp(a) has proatherogenic and prothrombotic properties and, is considered a risk factor for the development of cardiovascular disease (CVD) . Also, EDespite well-established evidence on Lp(a) as a risk factor of CVD, findings on relation between this marker and mortality are controversial. Also, this association has been studied more extensively in the general/healthy population than in patients. The Emerging Risk Factors Collaboration in 2009 investigated the association of Lp(a) concentration and risk of coronary heart disease, stroke, and mortality in healthy and general population. The findings of this individual participant meta-analysis of cohort and nested case-control studies indicated the risk ratios (RR) of 1.14 : 1.07–1.22, number of studies (n)= 24) and 1.01  for aggregate of coronary and nonvascular mortality . On the Regardless of a considerable number of publications, few studies have summarized the association of circulating Lp(a) with the risk of mortality. The latest meta-analysis on this association was published in 2011, which investigated the association with all-cause mortality as a secondary outcome in the general population . AnotherThis systematic review was designed and reported in accordance with PRISMA guidelines . The stuSupplementary Table-1. To complete our search, the references of the included studies and published reviews were manually reviewed.Embase.com, Medline ALL (Ovid), Web of science Core Collection, and Cochrane Central were searched until June 8, 2021. Lipoprotein(a) and mortality-related keywords were combined in the search strategy. Furthermore, filters to exclude conference abstracts, case-report studies, non-adult populations, and animal studies were applied. The search strategy was developed by an expert research librarian (WB). The search results were imported in EndNote and de-duplicated with the method as described by Bramer et al. ,,Suppleme studies , 25 8]  SupplemeSupplementary Table-2 presents the main characteristics of the included studies in the meta-analysis. Briefly, these studies have been published between 1998 and 2020. The association of Lp(a) concentration and all-cause or cause-specific mortality was reported in general or healthy populations [ulations , 58–70, ulations , 71–84, ulations , 31–33, ulations , 85, 86.ulations  and two ulations , 68. TheSupplementary Table-3.Additionally, 20 studies were graded as fair and the remaining as good quality, according to Newcastle – Ottawa quality assessment scale, as presented in 2: 75.34%, n = 19, Supplementary Figure-1) and in patients with CVD . No significant association was found in patients with CRF or DM. The results of meta-analysis were not robust in leave-one-out analysis in general population by removing several effect sizes [Supplementary Figure-6-A and B). Funnel plots showed no asymmetry in both populations .The association of Lp(a) and all-cause mortality is summarized in Table ct sizes , 65–67, 2: 54.6%, n = 14, Fig. 2: 61.7%, n = 4, Fig. nonlinearity > 0.05, Supplementary Figure-8-A and B, Supplementary Table-6).Moreover, there was no significant linear dose-response association either in general population , patients with CVD  and patients with DM . These results were robust in leave-one-out analysis  and no asymmetry was detected in funnel plots .Table 2: 0.0%) compared to the other regions .According to the findings of subgroups analyses in patients with CVD, higher levels of Lp(a) was associated with a higher risk of mortality in Asian studies . No non-linear dose response association was observed .In general population and patients with CVD, linear dose-response analysis revealed a higher risk of CVD-death with higher levels of Lp(a). A 50 mg/dL increase in Lp(a) concentration was associated with 31% and 15% greater risk of CVD-death in both general population and patients with CVD . Also, no significant linear .We found no significant association between Lp(a) concentration and non-CVD-death in the general population  and all-cause and cause specific mortality in the general population and in patients with chronic diseases.The association of Lp(a) and mortality might be explained by some pathways. Several meta-analyses have reported the association between Lp(a) and cardiovascular disease/events , 87. In The significant increase of the Lp(a)-associated risk for CVD death in the general population in our study is in line with the results of a study conducted by the Emerging Risk Factors Collaboration including 72,683 individuals from 24 studies, found that a one standard deviation (SD) (3.5 fold) higher Lp(a) was associated with 14% increase in CVD-death, while no significant association was observed in risk of non-vascular mortality . SimilarThis study has some limitations that should be considered while interpreting its results. Although we performed our analysis separately for studies in the general population, some studies in this population might not exclude patients with pre-existing diseases. Thus, this act may provide bias induced by pre-existing diseases. Residual confounding could have affected the results since we performed a meta-analysis of observational studies. However, there was little indication of heterogeneity between subgroups that adjusted for BMI or lipid-lowering medication, we could not perform subgroup analysis based on other confounding variables including hypertension status, blood lipid markers, physical activity and smoking as the majority of included studies presented these variables in their multivariable statistical models. Since most studies lacked multiple Lp(a) assessments, the reported relative risks in the original studies may have been subject to underestimation due to regression dilution bias and random measurement error . We obseThe present study also has several strengths. Unlike the majority of systematic reviews and meta-analyses that only focused on the associations in the general population, we present all available evidence relating to the association of Lp(a) and mortality irrespective of health condition. Additionally, we systematically summarized the available literature presenting the information on the association between Lp(a) and mortality in univariate models or using descriptive/frequency results in our supplementary materials. The linear and non-linear dose-response analyses enabled us to clarify the shape of the dose-response relations and to provide clear insight into the quantitative evaluations of the associations. To identify sources of heterogeneity, we performed various subgroup analysis based on geographical region, follow-up duration, whether studies were conducted in multiple cohorts or a single cohort, adjustment for main confounders, and Lp(a) assessment methods. We transformed the relative risk estimates which were often reported differently by each study to top vs. bottom third of baseline Lp(a) distribution to enable a consistent approach for running the meta-analysis.According to our findings, we have several suggestions for future studies. Research has indicated that Lp(a) concentrations are mainly determined by the number of KIV2 repeats genetically which inversely correlate with Lp(a) levels  and subsIn conclusion, our meta-analysis suggests higher risks of all-cause and CVD mortality in both general population  and in patients with CVD  in third versus first tertile of Lp(a) distribution. Also, the linear dose-response analysis showed a higher risk of CVD mortality in both patients with CVD (HR: 1.15) and in the general population (HR: 1.31) per 50 mg/dL increase in Lp(a) concentration. According to 2019 ESC/EAS Guidelines, Lp(a) should be measured at least once in each adult person’s lifetime to identify those who have very high inherited Lp(a) levels > 180 mg/dL, who may have a lifetime risk of atherosclerotic cardiovascular disease which is approximately equivalent to the risk associated with heterozygous familial hypercholesterolemia . Other pBelow is the link to the electronic supplementary material.Supplementary Material 1


Pnonlinearity = 0.522, Pdose-response = 0.005). Overall, among the dietary sources of protein, the risk of IBD increased only with increasing total meat intake, and the consumption of protein from dairy products was found to be a protective factor against the IBD risk.We aimed to conduct this dose-dependent meta-analysis to examine the relation between total protein, animal protein, and its sources with inflammatory bowel disease (IBD). We searched databases, comprising PubMed/Medline, Web of Science (ISI), Embase, and Google Scholar, for the published studies up to 28 March 2023. Prospective cohort study designs that investigated associations between dietary intake of various animal protein sources and with risk of IBD in the general population were identified. Eleven prospective cohort studies with 4,302,554 participants and 8067 cases were considered eligible. Findings indicated that higher intake of dairy was significantly associated with a lower risk of IBD , Crohn disease , and ulcerative colitis . There was no association between different sources of animal protein and the risk of IBD. The dose-response analysis suggested that each 100 g/d increment in dietary total meat consumption was associated with a 38% greater risk of IBD. Moreover, a positive linear association was found between total meat intake and risk of IBD (This trial was registered at PROSPERO as CRD42023397719. To our knowledge, this is the first dose-response meta-analysis of prospective cohort studies to assess the relationship between the amount of consumed dietary total protein, animal protein, and animal protein sources and the risk of developing inflammatory bowel disease in the adult population.Inflammatory bowel disease (IBD) is specified by recurrent inflammation of the intestinal tract that needs life-long supervision . Crohn dDysregulated immune response, genetic susceptibility, gut microbiome, and environmental factors may play a part in IBD development, but they failed to indicate sufficient information regarding dietary triggers of IBD ,9. GenerNarula et al. , in a reThese controversial results might be due to the fact that previous relevant reviews have mostly concentrated on results from case-control studies ,11,17 orWe pursued the frameworks specified in the Cochrane Handbook for Systematic Reviews to implement this meta-analysis . BesidesWe searched online databases, comprising Web of Science (ISI), PubMed/Medline, Embase, and Google Scholar, for the published studies up to 28 March 2023. A literature search was performed and expanded (S.T.), and 2 investigators (M.R. and F.N.) screened the titles/abstracts. The same 2 investigators separately evaluated the relevant full texts for eligibility. Discrepancies were addressed by consensus. Moreover, the references of published observational reviews on the relation between total protein, as well as animal protein, and its sources with the incidence of IBD were manually screened. We did not enforce any publication time or language restrictions. Our comprehensive search strategy is recorded in 1) included general adult study participants (aged ≥18 y); 2) utilized prospective cohort design; 3) reported exposure as the consumption of total protein, as well as animal protein, and its sources  in >2 categories; 4) reported the outcome as the incidence rate of IBD (CD or UC); and 5) reported suitable effect estimates, including RR, HR, or OR and their 95% CIs.Two of the investigators (S.T. and M.R.) reviewed the title/abstracts of each article discovered in the literature search to determine articles that: If more than one cohort study indicated similar data, we selected the ones with thorough records for dose-dependent analysis . Other than that, the articles with the largest sample sizes were chosen. We excluded articles with cross-sectional or case-control study designs, as well as publications conducted among patients with all gastrointestinal disorders other than IBD (CD or UC), such as celiac disease, gastroesophageal reflux disease, irritable bowel syndrome, small intestinal bacterial overgrowth, gallstones, and so on.The relevant data extracted from all studies by 2 independent investigators (S.T. and M.R.) were outlined as follows: the first author, cohort name, location of study, year of publication, confounders adjusted, follow-up length, instruments used to assess total protein, as well as animal protein, and its sources, number of participants/cases, mean age, method of recognizing outcome, type of exposure, comparison categories and relevant effect sizes  with their 95% CIs. Any discrepancies were addressed by a discussion with the third investigator (H.M.).We performed a quality assessment of cohort studies applying the Risk Of Bias In Non-randomized Studies-of Interventions tool . EvaluatQ test and I2 statistic [We chose the RRs with 95% CI as the effect estimate for our meta-analysis. The reported HRs were deemed equal to RRs . For cohtatistic .We did a subgroup analysis according to the region, sex, length of study duration, number of overall participants, and subjects with IBD, dietary assessment, case ascertainment (method of recognizing patients with IBD), and adjusting for confounders, comprising physical activity, sex, BMI, smoking status, EI, and consumption of alcohol. We also conducted meta-regression analyses to discover the potential source of heterogeneity. Publication bias was estimated by Egger’s test and Begg’s test . We carrFor linear dose-dependent analysis, the method developed by Greenland and Longnecker  and the To conduct our analysis, we deemed the median of each protein category, the numbers of subject or a person year, the number of subjects with IBD, and adjusted effect sizes within a minimum of 2 classes of exposures were pooled from every study. Moreover, most studies ranked protein intake as tertiles, quartiles, or quintiles; as a result, the highest protein intake category  compared with the lowest category was compiled for analysis. For cohort studies that documented effect estimates per specific increment in the quantity of exposure, the log effect estimate was exponentiated by multiplying the cohort study-certain consumption of the exposure to get the effect estimate for 1 extra serving of exposure ,29. WhenP value of <0.05 was regarded to be significant.In addition, we estimated curve linear or nonlinear dose-dependent relation between total protein, as well as animal protein, and its sources and IBD incidence by applying the restricted cubic line that was made up of 3 nodes at the fixed percentiles  . The corRating the evidence was performed by applying the updated Grading of Recommendations, Assessment, Development, and Evaluations (GRADE) framework ,34. GRADA total of 3200 records were identified from the initial search . Among these, 445 publications were duplicated, animal studies and nonoriginal articles. After title and abstract screening, 2704 irrelevant documents were removed. Consequently, by evaluating the full text, 40 articles were further excluded for various reasons, as shown in All population-based cohort studies were carried out in the general adult population. These cohort studies were dated from 2010  to 2022 Nine cohort studies were judged at moderate risk of bias , whereasn = 4), CD , and UC   examined the association between dietary total protein intake and risk of IBD ,37,38,44; n = 3) . Moreove; n = 3) . There w; n = 3) A.TABLE 1n = 565,974) were included in this association [n = 3), 1.34 , and 1.20 , respectively , CD , and UC . In the subgroup analyses, geographic locations and those that controlled for physical activity were identified as potential sources of heterogeneity  with 6824 cases were considered in the analysis of the highest and lowest categories of processed meat intake and risk of IBD [n = 5), CD , and UC   .n = 2), CD , and UC   .A linear dose-response investigated that each 100 g/d increase in dietary poultry intake was not associated with the risk of IBD  with 4757 cases evaluated the relation between dietary fish intake and risk of IBD [n = 5), CD , and UC   .Each 15 g/d increment in dietary fish intake was not associated with the risk of IBD , CD , and UC   .Findings indicated that each 100 g/d increment in dietary total meat consumption was associated with a 38% greater risk of IBD , CD , and UC   were included in the analysis of dairy intake . Compari; n = 5) . However; n = 5) .The dose-response analysis revealed that each additional 200 g of dairy consumption per day was not significantly associated with a lower risk of IBD , CD , and UC   .No linear association between dietary egg intake and risk of IBD was found  in comparison with the lowest protein intake category was compiled for analysis. This is a routine method; however, it is relatively restrictive because it cannot determine the optimum dose required for decision making. Thus, we further evaluated our data with dose-response analysis, which could help us better understand whether the risk of IBD was affected by increased intake of protein. However, this study had some limitations that should be considered in the interpretation of the data. First, retrospective assessments of dietary or other factors may have resulted in recall bias. Second, high heterogeneity was found in some of the investigated factors. Third, some of the evaluated studies had moderate and low quality from a methodological point of view.In conclusion, the results showed no significant relationship between the consumption of most dietary protein sources and the incidence of IBD. Among the dietary sources of proteins, the risk of IBD increased only with increasing total meat intake, and the consumption of protein from dairy products was found to be a protective factor against the IBD risk. More studies, especially intervention trials, are needed to confirm the results of this study.Tehran University of Medical Sciences’ Students’ Scientific Research Center (SSRC) (code: 1401-4-125-64978).Supported by The authors report no conflicts of interest.