By examining evidence from 30 non-randomized studies, we found that thiazide diuretic users have a higher risk of skin cancer than non-thiazide diuretic users. A small but consistent risk was identified across all types of skin cancers, including the more dangerous malignant melanoma  and non-melanoma skin cancers . The high cancer risk associated with thiazides, especially the risk of malignant melanoma, is concerning. Individual skin cancer risk assessment, monitoring, and consideration of suitable drug alternatives are needed before the long-term use of these agents. Our findings suggest that to promote the rational use of anti-hypertensive medication, the benefits of thiazide diuretics must be weighed against potential safety concerns in terms of skin cancer risks.n = 10,129,196; pooled adjusted OR, 1.10; 95% CI, 1.04–1.15; p < 0.001; strength of evidence, very low; very small harmful effect), BCC , and SCC  than non-users. Thiazide diuretic use was also associated with a higher risk of lip cancer , whereas other secondary outcomes were inconclusive. Conclusions: Thiazide diuretics are associated with the risk of all skin cancer types, including malignant melanoma; thus, they should be used with caution in clinical practice.Background: The use of thiazide diuretics is associated with skin cancer risk; however, whether this applies to all skin cancer types is unclear. Methods: In this meta-analysis, we searched multiple electronic databases and gray literature up to 10 April 2022, with no language restrictions, to identify relevant randomized controlled trials (RCTs) and non-randomized studies  that investigated the association between thiazide diuretics and skin cancer. The primary outcomes of interest were malignant melanoma and non-melanoma skin cancer . Secondary outcomes included other skin cancers . We used a random-effects meta-analysis to estimate pooled adjusted odds ratios (ORs) and 95% confidence intervals (CIs). Results: Thirty non-randomized studies  were included. Thiazide diuretic users had a higher risk of malignant melanoma (17 studies;  Numerous classes of anti-hypertensive medications are available to treat high blood pressure. Thiazide diuretics are commonly prescribed anti-hypertensive medications, accounting for approximately 30% of all prescriptions in the United States and Western Europe ,2. OwingEmerging evidence from post-marketing surveillance studies suggests that thiazide diuretics can increase the risk of skin cancer . In the Among the skin cancers, melanomas are the deadliest. They exhibit an early metastatic potential and are often highly aggressive, though whether the risk associated with thiazides differs according to skin cancer types is unclear ,9,10,11.This meta-analysis followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines  and the We searched electronic databases including Medline, Embase, PubMed, Cochrane Library, Web of Science, Scopus, and CINAHL from inception to 7 May 2021, with no language restrictions. Combinations of Medical Subject Headings and search terms were used, including pharmacological class or individual drugs , and skin cancers . The full search strategy for each database is provided in We included both RCTs and non-randomized studies  that (i) investigated thiazide diuretic use for any indication and the risk of skin cancer among individuals aged 12 years or older; (ii) consisted of two or more groups, with one group representing users of thiazide or thiazide-type diuretics ; and (iii) reported the occurrence of any type of skin cancer. We excluded studies that (i) reported only unadjusted effect estimates or lacked information to calculate risk estimates; (ii) had a small sample size (less than 50 patients); and (iii) were case series/case reports, N-of-one trials, pharmacokinetic/pharmacodynamic studies, cross-sectional studies, RCTs, and reviews. Details of the selection criteria are described in The primary outcomes of interest were the major skin cancer types: (i) malignant melanoma and specific subtypes ; and (ii) non-melanoma skin cancer . The secondary outcomes were other forms of skin cancer .Initially, two investigators  independently screened eligible articles based on the titles and abstracts of records identified through systematic searches. Thereafter, a full-text review was conducted to identify the final set of studies for inclusion. Potentially eligible articles published in languages other than English were translated before full-text assessment. Any discrepancies in study selection at either stage were resolved through a team discussion.Using a standardized approach, two investigators  independently extracted information about (i) study characteristics, including study design , sample size, study population and setting, study period, statistical analysis methods (multivariable or propensity score approach), and risk factors adjusted for when deriving effect estimates; (ii) patient characteristics ; and (iii) specific exposure and control groups, and predefined outcomes of interest . For studies with incomplete or unclear information, the first or corresponding authors were contacted for clarification. If the authors did not reply after two attempts, we excluded their study from the meta-analyses. The final data set was independently cross-checked by two investigators  to resolve any discrepancies.Two investigators  independently assessed the methodological quality of each study using the Cochrane risk of bias in randomized trials (RoB 2)  and NewcAll analyses were performed, and forest plots were created using Stata software . We used adjusted odds ratios (ORs) with the greatest degree of confounder adjustment in meta-analyses assessing the association between thiazide diuretic use and the outcomes of interest. Since the methodological approach varied across included studies, we used the random-effects model to estimate pooled adjusted ORs with 95% confidence intervals (CIs) to address heterogeneity across all included studies . Using tQ test, with a p-value of less than 0.100 indicating significant heterogeneity. The 2I index and tau-squared (τ2) statistics were also used to classify the degree of heterogeneity as low , moderate , or high  ; n = 19780476; pooled adjusted OR, 1.05; 95% CI, 1.02–1.09; p = 0.003; high heterogeneity [I2 = 87.2%]; n = 16387862; pooled adjusted OR, 1.35; 95% CI, 1.22–1.48; p < 0.001; high heterogeneity [I2 = 97.1%]; n = 5668737; pooled adjusted OR, 1.08; 95% CI, 1.03–1.12; p = 0.001; high heterogeneity [I2 = 83·0%]; n = 161491; pooled adjusted OR, 1.92; 95% CI, 1.52–2.42; p < 0.001; moderate heterogeneity [I2 = 51.5%]; The summary findings, strength of evidence, and conclusion on the association between thiazide diuretic use and risk of all skin cancer types are shown in A priori subgroup analyses based on participant characteristics, secondary outcomes, and dose- and duration-response relationships could not be performed because of limited details on diuretic use, skin conditions, and ultraviolet exposure. However, the risk among thiazide diuretic users appeared in individuals who took hydrochlorothiazide, studies with a sample size of more than 10,000, case-control studies, and studies conducted in Europe and North America .The International Agency for Research on Cancer has classified hydrochlorothiazide as a possible carcinogenic medication (group 2B) . SubsequConcerning previous meta-analyses of non-randomized studies, the study by Gandini et al. , which in = 58,185) comparing thiazide diuretics with other anti-hypertensive medications, with trial durations of 4.5 years . They found that thiazide diuretic use was associated with an increased risk of cancer death ; however, no evidence of association was recognized in the incidence of unspecified type of skin cancer. In fact, apart from outcomes with delayed onsets, such as cancer-related treatment interventions, RCTs are generally not intended to assess the long-term carcinogenic safety profiles of anti-hypertensive medications. Meanwhile, our study has an important methodological contribution because we included real-world evidence from a large sample with diverse participant characteristics and a long follow-up time, which allowed us to quantify the risk of all types of skin cancers associated with thiazide diuretic use in clinical practice. Interestingly, the risk of skin cancer associated with thiazide diuretics seemed to be limited to Europe and North America, as the association was not observed in other regions (Taiwan or Korea). We postulated that Europeans and North Americans, who are mostly fair-skinned, are more susceptible to ultraviolet radiation and the photosensitizing properties of thiazide diuretics. This finding is supported by a systematic review by Lopes et al. [In 2021, Copland et al.  performes et al.  which inOur study included up-to-date and expanded evidence on the association between thiazide diuretic use and the risk of all skin cancer types. From a methodological viewpoint, we performed extensive searching without language restrictions along with a rigorous and comprehensive systematic review to draw evidence-based conclusions. Moreover, the results were robust based on sensitivity analyses and no evidence of publication bias.Nevertheless, our meta-analysis had several limitations. First, since our results are based on non-randomized studies, the risk of selection bias in terms of confounding by indication or contraindication should be considered when interpreting our findings. Therefore, conclusions about causality relationships cannot be drawn. Second, apart from the nature of non-randomized studies, we found that uncertainty and potential unmeasured confounders existed based on the prediction interval and E-value, respectively. Consequently, we downgraded and judged the strength of evidence to be very low. Third, in the risk-of-bias assessment, only 20 studies (66.7%) were of high quality (NOS of 8 or greater). However, a post-hoc sensitivity analysis, including studies with the highest quality, showed no substantial difference from the main findings. Fourth, our findings may have information bias because most of the included studies relied on electronic databases or routinely collected administrative data. Nonetheless, most studies identified skin cancer cases using histological confirmation or through regional or national cancer registries. Fifth, differences were observed in the study population and exposure to individual thiazide diuretics, which could explain the moderate or high heterogeneity of the pooled effect estimates. Moreover, the results were incompletely adjusted for known risk factors, namely ultraviolet radiation exposure. Different skin phenotypes such as fair or colored skin were addressed only in a few included studies. Lastly, details on individual thiazide diuretic use and key participant characteristics  are lacking; thus, risk estimates in subpopulations cannot be derived.Given the limited effect size and evidence certainty, our findings provide the best available evidence on the carcinogenic safety of thiazide diuretics in general practice. Although thiazide diuretic use only slightly increases the risk of skin cancers, our study underscores that the benefits and risks of this medication class should be balanced, especially with long-term use in the management of chronic conditions such as hypertension. Proactive monitoring is warranted in thiazide diuretic users, and those at risk of developing skin cancer should be identified. These include individuals with extreme exposure to ultraviolet radiation from sunlight, indoor tanning, or other artificial sources for medical or cosmetic purposes; fair skin and advanced age; a history of skin cancer or precancerous skin conditions; and immunocompromised conditions or use of immunosuppressive agents that may further increase the skin cancer risk ,64,65.To minimize skin cancer risk and promote rational drug use, physicians should encourage patients to use sunscreen and sun protection in an outdoor setting and reduce the use of indoor tanning or other artificial radiation sources, as well as provide information about sun safety when prescribing thiazide diuretics. Further collaborative longitudinal pharmacoepidemiological surveillance studies using real-world data are needed to confirm the causal association between thiazide diuretic use and skin cancer risk. Moreover, proactive screening studies and intervention trials focusing on skin cancer prevention strategies are warranted.With respect to the very low strength of evidence certainty, thiazide diuretics are associated with the risk of skin cancer, including malignant melanoma and non-melanoma skin cancer. Future studies examining other forms of skin cancers or individual thiazide diuretics, as well as dose- and duration-response relationships, are required. Our findings suggest that individual skin cancer risk assessments, consideration of suitable alternative medications, and limited duration of thiazide diuretic use may be potential mitigation strategies.


In this systematic review, we aimed to investigate the efficacy and safety of adding low-molecular-weight heparin (LMWH) or unfractionated heparin to low-dose aspirin (LDA) started ≤16 weeks'gestation in the prevention of preeclampsia (PE) in high-risk women.ClinicalTrials.gov databases were searched from their inception to April 2022 for randomized controlled trials (RCTs) that to determine whether the combined treatment of LMWH and LDA is better than single anticoagulant drugs in preventing PE and improving live birth rate of fetus in high-risk women with pregnancy ≤16 weeks. We also searched Embase, OVID MEDLINE and OVID MEDLINE in-process using the OVID platform.PubMed, Cochrane Library, Embase, and P < 0.05), small-for-gestational age , fetal and neonatal death  and gestational hypertension . It is worth emphasizing that LMWH (or unfractionated heparin) combined with LDA did not increase the risk of bleeding.14 RCTs involving 1,966 women were found. The LMWH (or unfractionated heparin) and LDA groups included 1,165 wemen, and the LDA group included 960 women. The meta-analysis showed that the addition of LMWH to LDA reduced the risk of PE (RR: 0.59, 95% CI: 0.44-0.79, LMWH combined with LDA can effectively improve the pregnancy outcome of women with high risk factors for PE and its complications. Although this study showed that combined medication also did not increase the risk of bleeding, but such results lack the support of large sample size studies. The clinical safety analysis of LMWH combined with LDA in patients with PE should be more carried out. Preeclampsia (PE) is characterized by hypertension and proteinuria, and it is one of the important factors leading to maternal and perinatal death. The incidence rate of all pregnant women is about 2–8% , 2. The At present, there are evidence-based evidences that the application of aspirin in high-risk groups of PE can effectively prevent the onset of PE and reduce adverse pregnancy outcomes , 9. StudIn recent years, some studies have reported the effect and safety of aspirin combined with LMWH in the prevention of PE and its complications. In this study, meta-analysis method was used to conduct a comprehensive evaluation of these related studies to make up for the small sample size of a single study. The primary purpose is to determine whether the combined treatment of LMWH and aspirin is better than single anticoagulant drugs in preventing PE and its complications in high-risk women with pregnancy ≤16 weeks.We performed a meta-analysis and wrote the article by conforming to the requirements illustrated in the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement . We compClinicalTrials.gov databases were searched from their inception to April 2022. We also searched Embase, OVID MEDLINE and OVID MEDLINE in-process using the OVID platform. The search terms that we used in the Pubmed database were as follows: #1 “pre-eclampsia” [title/abstract] OR “preeclampsia” [title/abstract] OR “PE” [title/abstract] OR “eclampsia” [title/abstract]; #2 “heparin” [title/abstract] OR “Low-Molecular-Weightheparin” [title/abstract] OR “LMWH” [title/abstract]; #3 “aspirin” [title/abstract]; #4 #1 AND #2 AND #3. The databases were searched for published studies in English, including systematic reviews and meta-analysis, that were related to the treatment with LMWH or heparin (with LDA) in women at high risk of PE. After the search is complete, the documents were exported in full text format.PubMed, Cochrane Library, Embase, and Inclusion criteria were: (1) randomized controlled trials (RCTs) studies comparing LDA with LMWH or heparin for prevention of PE; (2) studies including women who had any known high risk factors for PE and its complications, such as adverse obstetric history of previous PE, small-for-gestational age (SGA) or FGR, placental abruption, and medical history including thrombophilia, autoimmune disease, and other chronic diseases; (3) published paper.Animal experiments and non-RCTs were excluded from the meta-analysis. In addition, studies were excluded if they included other treatment options.The primary outcomes were PE and live birth rate, with secondary outcomes including placental abruption, severe PE (sPE), gestational hypertension, FGR or SGA, fetal and neonatal death and HELLP  syndrome. We also analyzed the presence of adverse events such as intrapartum or postpartum hemorrhage.In this context, it is recommended that women are diagnosed with PE if they present with the following severe features: systolic blood pressure of 140 mmHg or more or diastolic blood pressure of 90 mmHg or more on two occasions at least 4 h apart after 20weeks of gestation in a woman with a previously normal blood pressure plus proteinuria, defined asurinary excretion of 300 mg or more per 24 h urine collection (or this amount extrapolated from a timed collection)orc Protein/creatinine ratio of 0.3 mg/dL or more orc Dipstick reading of 2+(used only if other quantitativemethods not available) .Two authors (BX and YY) independently screened and retrieved the relevant literatures. Then, after reading the title and abstract of the article, they excluded the studies that did not meet the inclusion criteria. Full texts of final selected articles and data extraction were performed independently by other 2 researchers (LZ and YW), and any discrepancies were resolved by discussion.The extracted data included names of the first authors of selected articles, basic information on studies , characteristics of patients , and reported outcomes .The risk of bias of RCTs was assessed using the Cochrane Collaboration's tool  based onI2 to quantitatively judge the extent of heterogeneity. The fixed-effects model was applied at P > 0.1 and I2 < 40%, which showed that there was little heterogeneity between studies. Otherwise, the random-effects model was used for analysis  with 95% confidence interval (CI) between the experimental and control groups. For dichotomous outcome data, the risk ratio (RR) or odds ratio (OR) with 95% CI was calculated. The chi-square test was used to assess heterogeneity between studies, and combined with analysis .Begg's and Egger's tests of asymmetry were performed using the statistical software Stata 12.0 to assess potential publication bias. The tests were also used to identify outliers .Thousand six hundred and seventy eight articles were obtained from the search through the above mentioned databases. Of these, 1,146 articles were removed due to record duplication and 447 studies were excluded based on their titles and abstracts. Finally, a total of 14 studies that included 1,966 women were considered in the systematic review and meta-analysis based on the inclusion and exclusion criteria .The baseline characteristics of the included studies were presented in the The Cochrane Collaboration's Risk of Bias tool  was usedThe highest risk of bias occurred in the blinding of participants and personnel. This is because neither study personnel nor participants in five studies , 26, 29 Among all the included studies, six studies , 26, 30 n = 1,452), the pooled results obtained using a fixed-effects model showed that the addition of LMWH to LDA reduced the risk of PE . In the subgroup analysis, the addition of LMWH to LDA reduced the risk of PE in women with a history of PE  and miscarriages , respectively   .n = 1,464), the pooled results obtained using a fixed-effects model showed that the addition of LMWH to LDA did not reduced the risk of placental abruption . In the subgroup analysis, the addition of LMWH to LDA did not reduced the risk of placental abruptionin women with a history of PE . Conversely, in women with a history of miscarriages, the addition of LMWH to LDA can reduce the risk of placental abruption   .n = 889) reported the SGA, the pooled results obtained using a fixed-effects model showed that the addition of LMWH to LDA reduced the the risk of SGA . But in the subgroup analysis, the addition of LMWH to LDA did not reduced the risk of SGA in women with a history of PE  and miscarriages , respectively  reported the sPE, the pooled results obtained using a random-effects model showed that the addition of LMWH to LDA did not reduced the the risk of sPE . In the subgroup analysis, the addition of LMWH to LDA did not reduced the risk of sPE in women with a history of PE . Conversely, in women with a history of miscarriages, the addition of LMWH to LDA can reduce the risk of sPE   .n = 414) reported the gestational hypertension, the pooled results obtained using a fixed-effects model showed that the addition of LMWH to LDA reduced the the risk of gestational hypertension . In the subgroup analysis, the addition of LMWH to LDA did not reduced the risk of gestational hypertension in women with a history of PE . However, in women with a history of miscarriages, the addition of LMWH to LDA can reduce the risk of gestational hypertension  (Three studies –21 (n =  = 0.05) .n = 491) reported the FGR, the pooled results obtained using a fixed-effects model showed that the addition of LMWH to LDA did not reduced the the risk of FGR . In the subgroup analysis, the addition of LMWH to LDA also did not reduced the risk of FGR in women with a history of PE  and miscarriages , respectively  reported the fetal and neonatal death, the pooled results obtained using a fixed-effects model showed that the addition of LMWH to LDA reduced the risk of fetal and neonatal death . In the subgroup analysis, the addition of LMWH to LDA also can reduced the risk of fetal and neonatal deathin women with a history of PE , but the addition of LMWH to LDA did not reduce the risk of fetal and neonatal death in women with a history of miscarriages   .n = 918) reported the HELLP syndrome, the pooled results obtained using a fixed-effects model showed that the addition of LMWH to LDA did not obviously reduced the risk of HELLP syndrome . In the subgroup analysis, the addition of LMWH to LDA also did not reduced the risk of HELLP syndrome in women with a history of PE  and miscarriages , respectively. Although the results were not statistically different, it can be seen from the Six studies , 26, 28 n = 471) reported the intrapartum or postpartum hemorrhage, the pooled results obtained using a fixed-effects model showed that the addition of LMWH to LDA did not increased the risk of intrapartum or postpartum hemorrhage   .P > 0.05  and cyclooxygenase 2 (COX-2), and finally inhibits platelet aggregation , therebyLMWH strengthens its inactivation of coagulation factors IIa, IXa and Xa by combining with antithrombin, thus reducing the generation of corresponding coagulation factors. It not only has the effects of anticoagulant, preventing placental thrombosis and placental infarction , but alsStudies have shown that LDA combined with LMWH may offset the adverse reactions caused by some drugs and increase the effect of the two drugs . For preThe strengths of our study include the inclusion of more RCT studies and more patients comparing the effects of combination and single drug than other meta-analysis. Although there may be heterogeneity in our results, it also proves to some extent that our results are applicable to women with high risk factors for PE and its complications. At the same time, we also evaluated the bleeding risk of combined drugs, which is also a key safety index to be considered in the process of clinical use.This study has some limitations. First, the quality grades of the included studies were inconsistent. Among them, five studies , 26, 29 Therefore, the efficacy of aspirin combined with LMWH is better than that of single drug, and can improve maternal and infant outcomes to a certain extent. It is worthy of clinical promotion, but it needs to be further verified by strictly designed, large sample and multi center randomized controlled trials.The results of this meta-analysis showed that LMWH combined with LDA can effectively improve the pregnancy outcome of women with high risk factors and reduce the incidence of PE, SGA, gestational hypertension, fetal and neonatal death. At the same time, there are three studies that report women's bleeding. The results show that the combination of drugs does not increase the risk of bleeding, but such results lack the support of large sample size studies. The clinical safety analysis of LMWH combined with LDA in patients with PE should be more carried out.The original contributions presented in the study are included in the article/supplementary material, further inquiries can be directed to the corresponding authors.LZ and YaW planned and designed the study. LZ, BX, YY, and YuW screened studies and performed the systematic review and extracted data from included studies. YY, BX, and LZ analyzed the data and performed the quality assessment of studies. LZ drafted the manuscript. All authors contributed to the discussion section of the manuscript. All authors contributed to the article and approved the submitted version.


Background: The relationship between CILP (1184T>C) and IL-1α(+889C/T) polymorphisms and intervertebral disc degeneration (IDD) have been explored in several studies but the results were conflicting. The aim of the study was to evaluate and synthesize the currently available data on the association between CILP (1184T>C) and IL-1α(+889C/T) polymorphisms and susceptibility of phenotype-dependent radiologic IDD (RIDD) and symptomatic intervertebral disk herniation (SIDH).Methods: A computerized literature search was in PubMed, Cochrane Library, Embase, China National Knowledge Infrastructure database, and Web of Science. The pooled results were presented as odds ratios (ORs) with 95% confidence intervals (CIs). Moreover, the false-positive report probability (FPRP) test and trial sequential analysis (TSA) were applied to estimate the significant results.Results: Our evidence demonstrated that IL-1α(+889C/T) was significant associated with RIDD  and SIDH . However, the results were not noteworthy under the FPRP test and TSA analysis. Additionally, CILP (1184T>C) polymorphism was significantly associated with RIDD with adequate evidence  instead of SIDH.Conclusion: The current meta-analysis illustrated firm evidence that CILP (1184T>C) polymorphism was significantly associated with the susceptibility of RIDD. However, the significant associations between IL-1α(+889C/T) and RIDD and SIDH were less credible. Thus, more multi-center studies with diverse populations were required to verify the results. Low back pain (LBP), one of the most common muscle-skeletal diseases, contributes to lower life quality and even disability. Over 80% of elderly people are afflicted with LBP to some extent during their lifetime . Intervevia the TGF-β1signaling pathway  contains a luxuriant extracellular matrix (ECM) such as proteoglycans and collagen. Accumulating evidence has highlighted the important role of ECM homeostasis play in the development of IDD. The imbalance of ECM homeostasis may cause diverse changes in disc including height reduction, dehydration, endplate alteration, etc. . Overexp pathway .Given the vital function of the IL-1α and CILP in ECM metabolism, numerous candidate polymorphism loci correlated to IDD predisposition were identified. Various studies indicated that CILP (1184T>C) and IL-1α(+889C/T) polymorphisms could be significantly correlated with the susceptibility of IDD . HoweverWe performed a literature retrieval of PubMed, China National Knowledge Infrastructure database, Web of Science, Cochrane Library database, and Embase by using the following terms  AND  AND (“IL-1α” or “CILP” or “IL-1A”). The bibliography of associated studies was also examined for additional correlative studies.Studies included in the meta-analysis should be based on the following criteria: 1) case-control study; 2) the consequence was the investigation of the correlation between CILP (1184T>C) IL-1α(+889C/T) or rs1800587 or rs2073711and IDD or SIDH; 3) genotype frequencies were available to calculate an odds ratio (OR) with 95% confidence intervals (CIs); 4) grade criteria or phenotypes of IDD were described in detail.The exclusion standards were listed as follows: 1) other types of study design; 2) genotype frequencies were not present in the articles; 3) studies contained overlapping data from previous research; 4) publication language was not English.The quality evaluation of each study was performed independently by two investigators (JCL and XQS) following the Newcastle-Ottawa Scale (NOS). Dissension between reviewers would be discussed to reach a consensus. Based on the NOS standard, only studies with calculating scores >5 were deemed of high methodological quality.Two reviewers independently extracted the following information on basis of a standard form: 1) Name of the First author 2) Publication information 3) Country 4) Ethnicity 5) Genotype frequencies 6) Numbers of cases and controls 7) Phenotypes of IDD.2 >50% or p < 0.05  and CILP (1184T>C)in different phenotypes of RIDD meanwhile stratified analysis was conducted in SIDH to test the genetic effect of CILP (1184T>C) in different ethnicity. False-positive report probability (FPRP) tests were performed to correct all significant pooled ORs. According to the calculating Excel spreadsheet  with a 5% type I error and 20% type II error (In order to control type I and type II errors and predict the required information size (RIS) in the research, the trial sequential analysis (TSA) was carried out in the TSA software . No statistical significance was identified regarding the recessive and heterozygous models  polymorphism and RIDD susceptibility. Obvious heterogeneity was detected in the recessive and heterozygous models. The pooled OR confirmed significant associations in three models   .p = 0.002), dominant model , and homozygous model   but no association with RIDD (Schneiderman grade) susceptibility (Four eligible studies comprising 897 cases and 1010 controls  are incl= 0.015)   . No signtibility .A total of 898 cases and 1111 controls  Figure . All modA ruling one out in turn strategy was adopted to estimate the effect of exclusion for each study on pooling ORs. The overall estimates did not alter under any genetic model when each individual study was omitted in turn, demonstrating the robustness of the findings . Begg’s We performed the FPRP test to investigate the false-positive results. The relationship between IL-1α(+889C/T) polymorphism and IDD and SIDH was not noteworthy (FPRP >0.2) in any model while the significant association between CILP (1184T>C) polymorphism and RIDD was considered a true positive result. The cumulative z-curve did not pass the TSA and RIS line for the association between IL-1α(+889C/T) polymorphism and IDD, indicating more studies are required . For IL-AS a multifactorial disease, IDD is thought to be mainly caused by overloading and senescence . In a brIn our meta-analysis, significant associations were detected between IL-1α rs1800587 polymorphism and IDD and SIDH but more evidence is required to confirm the conclusion. The subgroup analysis confirmed that IL-1α rs1800587 polymorphism was correlated with the risk of both RIDD (Solovieva grade) and SIDH but not RIDD (Modic change). Thus, we assumed that the change of expression due to rs1800587 polymorphism might mainly participate in the degeneration of NP instead of the endplate. Moreover, the homozygous model had the highest risk score. Compared with RIDD, a stronger association was detected between IL-1α(+889C/T) and SIDH. The results highlighted the importance of IL-1α polymorphism in the occurrence and progression of IDD.Moreover, the pooled ORs and 95% CIs and TSA result demonstrated that CILP (1184T>C) was a credible susceptibility variant for RIDD (Pfirrmann grade) instead of SIDH, indicating the genetic effect involved in the initial stage of NP degeneration. Consistent with the results of IL-1α(+889C/T), the highest risk score was detected in the homozygous model. Furthermore, CILP rs2073711 is associated with SIDH in an ethnicity-dependent manner, which was mainly due to the difference in allele frequencies for ethnicity. The C allele was found to be a protective factor for Caucasians but a risk factor for Asians. Additionally, the C allele is also a protective factor for knee arthritis in the European population , which iTo our best knowledge, this research is the first to comprehensively assess the genetic effect of IL-1α(+889C/T) and CILP (1184T>C) in different phenotypes of IDD. The significant associations were evaluated using the FPRP test and TSA. All eligible data was extracted to provide more convincing evidence. Our findings provided a novel perspective on the relationship between genetic polymorphisms of IL-1α and CILP and IDD phenotype-dependent susceptibility and emphasized its role in IDD early screening and SIDH prevention. Population with high-risk genotypes should avoid longtime overloading and other risk factors leading to the development of IDD.However, limitations should be claimed. First, despite a larger sample size in this study compared with previous studies, samples with other phenotypes of IDD are required to estimate the false-positive result. Second, this study did not estimate multiple confounding factors such as gender, occupation, and environmental factors, which might affect the effect estimates. Third, candidate gene studies have limitations in dealing with complex disorders and the results are often unrepeatable . A largeThe current meta-analysis illustrated adequate evidence that CILP (1184T>C) polymorphism was significantly associated with the susceptibility of RIDD. However, the significant associations between IL-1α(+889C/T) and RIDD and SIDH were not noteworthy. Further research with larger sample size and various populations is required to verify the outcome.


Dear editor,The pooled analysis in Figure 2 revealed a nonsignificant association between vitamin D level and COVID-19 risk and death. However, it is obvious that the pooled effect size might be influenced by a study conducted by Hastie et al. . This stP < 0.001), whereas Kaufmann had adjusted for race/ethnicity, gender and latitude [Confounding factors and methods of measuring vitamin D level: While using the adjusted odds ratio plays a key role in attenuating the effects of the confounding variables, the studies included differ considerably on the types of confounding adjustment. For example, the included odds ratio by Kanz et al.  were adjlatitude . Also, wlatitude , the poslatitude .In the statistical analysis section, the authors carried out the pooled analysis using a random effect model. In view of the few studies included in the quantitative analysis, the choice of which effect model to use (random versus fixed effects model) needs to be explained to the reader. While the random effects model can still be applied, the fixed effects model might be more suitable in this meta-analysis .I read with great interest the study entitled “Low vitamin D levels do not aggravate COVID-19 risk or death, and vitamin D supplementation does not improve outcomes in hospitalized patients with COVID-19: a meta-analysis and GRADE assessment of cohort studies and RCTs” by Chen et al. [Considering the above factors, the results of this analysis need to be cautiously interpreted, and well-designed prospective clinical trials are necessary.


Obstructive sleep apnea (OSA) as an independent cardiovascular risk factor has been proposed, but the mechanisms underlying cardiovascular disease is far from being completely elucidated. Leptin, an inflammatory cytokine produced by adipocytes, contributes to the modulation of metabolism, respiratory control, and inflammation, which are factors associated with cardiovascular disease. Serum levels of leptin in children with OSA have shown conflicting results in previous studies.We performed a meta-analysis to clarify the correlation between leptin expression of the OSA patients following the PRISMA. PubMed, Embase, and Web of Science were systematically searched for relevant studies, and then independently screened by two researchers, and analyzed the data through STATA version 12.0.P < .001), compared to the control group. Subgroup analysis were performed based on body mass index. The results of subgroup analysis demonstrated that the serum leptin concentration was correlated with body mass index in children with OSA .In a total of 5 articles including 469 participants, the data analysis showed that serum leptin levels were elevated in children with OSA (MD, 6.36; 95% CI, 0.24–12.49, The serum leptin levels were elevated in children with OSA, compared to the control group. It could add to our developing understanding of the pathogenesis and potential treatments for children with OSA, and help us to recognize the relevance of OSA in determining cardiovascular issues among children. Untreated OSA is associated with neurobehavioral problems, decreased attention, disturbed emotional regulation, decreased academic performance, nighttime enuresis, impaired growth and so on.,3 At last, OSA has a relevant impact on the cardiovascular system in children. Hence, early diagnosis of OSA may reduce the occurrence of systemic complications. At the same time, it is also important to explore the mechanisms of causing cardiovascular disease in children with OSA.Obstructive sleep apnea (OSA) is a disease that affect patients from infancy to adulthood. It affects 1% to 6% of all children and almost up to 59% of obese children. leptin, particularly in the context of hyperleptinemia, exerts detrimental effects in cardiovascular function and promotes adverse outcomes in cardiovascular disorders. Several studies reported the association between OSA and leptin.–9 Dalesio et al showed that leptin was significantly higher in the obese/OSA group than in the control or OSA-only group.10 The association between OSA and serum leptin levels is intricate and multidirectional since leptin levels can also be affected by obesity alone. It is reported that leptin may play roles in early diagnosis of OSA. Alteration in the levels of leptin is associated with the risk of cardiovascular diseases. However, contrary results was showed by Li et al that leptin levels were not different between OSA and non-OSA groups. The variability of serum leptin levels in children with OSA remains clearly inconclusive. In the present study, the databases of PubMed, Embase and Web of Science were searched for relevant publications and a meta-analysis was performed. We tried to clarify whether serum leptin level is elevated in children with OSA.Leptin is an adipocyte-derived hormone regulating energy expenditure and food intake and can be found in the circulation.We searched for English articles included in PubMed, Embase, and Web of Science. Search terms included the following key words: (obstructive sleep apnea hypopnea syndrome or sleep apnea or obstructive sleep apnea or obstructive sleep hypopnea or sleep-disordered breathing) and (adipokines or leptin) and (child or preschool children or teenager or adolescent). Potentially relevant articles were evaluated for inclusion against pre-specified eligibility and exclusion criteria.Eligible studies for the meta-analysis were required to meet the following inclusion criteria: participants at an age of 18 years or younger; serum leptin concentration was measured from morning fasting venous blood following overnight polysomnography; subjects received monitoring by polysomnography, and OSA was diagnosed if they had an obstructive apnea index ≥1. The exclusion criteria were the following: studies without sufficient data for meta-analysis; abstracts, reviews, letters, and case reports. The level of statistical significance was set at P < .05.Statistical analyses were performed by using STATA version 12.0. Weighted mean difference (WMD) and 95% confidence interval (CI) were used to present the statistical results for continuous outcomes. And an inverse variance method was used for continuous variables.2) value. P < .10 was considered statistical heterogeneity to be statistically significant. An I2 value above 75% indicated high heterogeneity, an I2 value between 50% and 75% moderate heterogeneity, and an I2 value between 25% and 50% low heterogeneity. A result was believed to be homogeneous when an I2 value was <25%. If I2 <50%, the study was believed to be homogeneous or to have low heterogeneity, and the fixed effects model was used to pool the results. If I2 >50%, the study was believed to be moderately or highly heterogeneous, and the random effect model was used to pool the data.,17Statistical heterogeneity was assessed on the basis of I square , and apnea-hypopnea index (AHI ≥10 vs <10). Sensitivity analysis was used to evaluate the stability of the meta-analysis results. Potential publication bias was evaluated by using funnel plot,Overall 40 relevant articles were extracted from the databases. And 19 articles were identified to be relevant by roughly screened in terms of abstract and title against inclusion and exclusion criteria. We retrieved the full texts of the articles and excluded several full texts for following reasons: 5 had no control or control group was selected as AHI ≥5 events/hour in children; two had no basic information; 2 records included patients someone older than 18 years; 5 studies unrelated to topic. The steps of the literature search are detailed in Figure ,13,20–22 The characteristics of the included studies were summarized, such as author, year, country, age, and so on. And the information of BMI, AHI and leptin of each study are given in Table Finally, five studies covering data from a total of 469 participants, were included in this meta-analysis.2 was 76%, indicating that the studies were high heterogeneous. Therefore, the random effects model was used to combine effect size. Meta-analysis exhibited that serum leptin levels in OSA group were 6.36 ng/mL higher than that in control group  Figure The value of IP = .04; Fig. P > .05) of the patients with OSA. Our analysis showed that the leptin level was not significantly correlated with the AHI of the patients All included studies reported the BMI values for each patient. Therefore, we performed a subgroup analysis of the articles according to obese or overweight. The total WMD in the studies with average BMI >25 was significant, with a corresponding value of 7.27  and Egger tests (P = .383) to assess publication bias in our study. The results were quite symmetric, indicating that the analysis did not include publication bias among the studies.We used Begg tests  which it was elevated more significantly in obese OSA patients. The overall heterogeneity was high in this meta-analysis (I2 = 76%); nevertheless, the sensitivity analysis indicated that no single study significantly influenced the pooled WMD, and Begg tests (P = .707) and Egger tests (P = .383) showed no publication bias in our study. Therefore, the result of our meta-analysis has relatively strong power. This is consistent with the previous findings that leptin levels appear to be determined by the degree of obesity as well as by the severity of OSA, particularly hypoxemia. The leptin is involved in energy expenditure, balance of blood glucose metabolism, inflammatory processes, and regulation of immune function. Animal studies demonstrated that leptin played key roles in respiratory control by acting centrally to alter ventilations. Moreover, leptin’s stimulation of oxidative stress, and at the cardiovascular level they result in vascular inflammation and vascular smooth muscle hypertrophy—all factors that contribute to atherosclerosis, hypertension, coronary heart disease and thrombosis. Sufficient clinical studies also showing that leptin levels are elevated in adult with OSA and associated with disease severity, age, and BMI.–25 However, the consistency of the results regarding the leptin levels of leptin in children with OSA remains to be studied. Many studies have reported that patients with OSA have higher leptin levels than control groups.,21,26,27 Both sleep deprivation and hypoxemia are thought to be critical causative factors in OSA-induced impact on leptin levels. Although, OSA and obesity are bidirectionally linked, One study demonstrated that serum leptin levels were significantly increases in children with OSA independently of obesity. Moreover, Treatment with continuous positive airway pressure (CPAP) and/or nasal corticosteroids in children with OSA led to a significant decrease in leptin while increases in leptin emerged in those with residual OSA.,29 A more attractive hypothesis from these studies would be that changes in the frequency of intermittent hypoxia might have changed the serum leptin. This supports our conclusion from another dimension.The present meta-analysis of 5 studies evaluated the serum leptin levels and compared these values between individuals with OSA and controls. The result showed that serum leptin levels were higher in children with OSA than in controls. And subgroup analysis revealed that the leptin level was significantly correlated with BMI ( No associations between OSA severity and the blood level of leptin in this clinical sample of overweight and obese children and adolescents. These results differ from the results of this meta-analysis maybe because differences in the included populations and small sample size.However, there are still some other sounds, such as leptin levels were not different between subjects with and without OSA.At last, the present meta-analysis contains several limitations. First, because of the differences in data statistics and experiment design, adequate data cannot be extracted from all or at least most of the included studies, numbers of articles were limited in the present meta-analysis. Second, obese children accounted for the majority of patients in the included studies which is associated with incidence of OSA in obese children. Third, the results may be heterogeneous due to the utilization of different methods and statistical analysis, sample size, and diagnostic work-up. Although, according to Begg’s funnel plots and the Egger’s test, the publication bias was not significant. So, according to present situation, additional studies with larger sample sizes are needed to acquire more representative and precise findings.In conclusion, the meta-analysis demonstrated that serum leptin levels were elevated in children with OSA, compared to the control group. It could add to our developing understanding of the pathogenesis and potential treatments for children with OSA, and help us to recognize the relevance of OSA in determining cardiovascular issues among children.Qing Cheng, designed study, collected data, analyzed data, wrote article, revised article; Xun Niu, designed study, collected data, wrote article, analyzed data, revised article; Yao He, designed study, wrote article, analyzed data, revised article; Liu-Qing Zhou, designed study, analyzed data, revised article; Yao Hu, designed study, revised article.Data curation: Yao Hu, Xun Niu.Formal analysis: Xun Niu.Software: Yao Hu, Qing Cheng.Writing – original draft: Yao He.Writing – review & editing: Liu-Qing Zhou, Qing Cheng.Qing Cheng’s name has been corrected from Qing Chen.


Opioid exposure during pregnancy has increased alarmingly in recent decades. However, the association between prenatal opioid exposure and congenital malformation risk has still been controversial. We aim to assess the association between opioid exposure during pregnancy and the risk of congenital malformations.PubMed, Embase, and Cochrane library of clinical trials were systematically searched to September 13th, 2021. Cohort studies reporting risk of congenital malformation after opioid exposure compared with non-exposure during pregnancy were included. Risk of studies was appraised with the ROBINS-I tool. Meta-analysis was conducted using the random-effects model. Subgroup analyses were conducted for the primary outcome based on indication, exposed period, whether adjusted data was used, and risk of bias assessment. Meta-regression was performed to evaluate the relation of publication year.Eighteen cohort studies with 7,077,709 patients were included. The results showed a significant increase in the risk of overall congenital malformation , major malformation , central nervous system malformation , and limb malformation  with opioid exposure during pregnancy. However, the predictive interval conveyed a different result on overall congenital malformation (95%PI: 0.82–2.09) and major malformation (95%PI: 0.82–2.09). No association between opioid exposure and overall congenital malformation in the first trimester  and prescribed for analgesic or antitussive treatment  were observed. In subgroups that study provided data adjusted for confounders  or identified moderate or serious risk of bias , no association was found.Opioid exposed in the first trimester or prescribed for analgesic or antitussive treatment did not increase the risk of overall congenital malformation. The findings should be discussed in caution considering the situation of individual patients and weigh out its potential risk of congenital malformation.CRD42021279445.Registration number: The online version contains supplementary material available at 10.1186/s12884-022-04733-9. Opioid is frequently prescribed for pain, such as lower back pain and pelvic joint pain, to reduce perception of pain during pregnancy. Also, opioid medications, such as methadone and buprenorphine, are used to treat opioid use disorder . Opioid The association between prenatal opioid exposure and congenital malformation risk has still been controversial. Two studies funded by the Centers for Disease Control and Prevention have set off an upsurge in studying the relationship between opioid use during pregnancy and congenital malformations , 14. SomThese findings call for the safety re-evaluation of opioid exposure during pregnancy to inform clinical practice. Therefore, we performed a meta-analysis using data from real-world cohort studies to assess the association between opioid exposure during pregnancy and the risk of congenital malformations.https://www.crd.york.ac.uk/prospero/ (registration number CRD42021279445) before searching articles.We followed the Meta-analysis of Observational Studies in Epidemiology  to perfoWe used PICOS model to select the population. The inclusion criteria were: (1) cohort studies; (2) investigated opioid use during pregnancy; (3) reported both opioids-exposed and -unexposed group; (4) reported on any congenital malformations and specific congenital malformation at birth; (5) reported available data, such as odds ratio (OR), adjusted OR, risk ratio (RR), adjusted RR, hazard ratio (HR), or data to calculate RR; (6) reported outcomes including any congenital malformations, major congenital malformation, and/or sub-categories of congenital malformations. The exclusion criteria were: (1) review, systematic review and meta-analysis, conference abstract, and case report; (2) not human studies; (3) did not clarify the exposure of opioid during pregnancy; (4) overlapped data source is included.We systematically searched PubMed, Embase, and Cochrane library of clinical trials up to September 13th, 2021. The search terms were attached in Table STwo reviewers  independently screened titles and abstracts through Endnote (version 9.3.2). Duplications were removed through Endnote and manually. We also screened the references lists of relevant reviews and articles. Any disagreement was resolved by discussion until consensus was reached or by consulting a third author (X.F).Data were independently extracted by two investigators (X.W and Y.W) for eligible studies. Disagreements were discussed and resolved by a third author (X.F). The data obtained for each study included first author, year of publication, study setting, drug used, exposure measurement, exposed period, outcome assessment, indication, sample size, congenital malformations with their risk estimates, 95% confidence interval (CI) and 95% prediction interval (PI). The primary outcome was overall congenital malformations. The secondary outcome was organ-specific congenital malformations.We assessed the risk of bias for each study included using the ROBINS-I tool , which iWe evaluated the level of evidence for each outcome using Grading of Recommendations, Assessment, Development and Evaluation (GRADE) approach . The res2 statistic. I2 > 50% or P <  0.05 was considered to indicate significant heterogeneity. We also addressed heterogeneity by calculating the 95% prediction interval for the pooled unadjusted OR, which gives an estimate of the point at which the true effects are to be expected for 95% of similar studies that might be conducted in the future [The meta-analysis was conducted with R (version 4.0.5). For the expected high heterogeneity in terms of the enrolled populations, DerSimonian and Laird random-effect models was used to pool RRs along with the corresponding 95% CIs. Due to the low prevalence of congenital malformation in the general population, we proposed RR, HR, and OR to be comparable. For those studies that did not report the RRs of congenital malformations, we used other risk measures, including ORs or HRs, as an approximation to the RRs. Therefore, we summarized them together using meta-analysis methods. The adjusted effect sizes were selected to pool the risk estimates preferentially. Statistical heterogeneity among studies was tested using Cochran’s Q test and the Ie future . The EggTo explore the sources of heterogeneity, subgroup analyses were run for primary outcome based on indication , exposed period , risk of bias assessment , and whether adjusted data was used . Due to the large span of publication year of included studies, we performed random-effects meta-regression analyses by the empirical Bayes method to estimate the between-study variance and the method by Hartung and Knapp was used to adjust statistics and evaluate the relation of covariates (year of publication) on the primary outcome. To evaluate the stability of the results, sensitivity analyses were performed with the leave-one-out method.One thousand one hundred seventeen studies were identified after database searching. 18 additional records were identified manually through references lists of relevant articles. After removing the duplications, 1030 studies were excluded by screening titles and abstracts. Only 18 studies , 27–39 wOpioid indications were analgesic or antitussive treatment, opioid abuse, and opioid abuse treatment. The total sample size of these studies ranged from 96 to 2,780,256. Kallen 2015 and Kallen 2013 were both from the Swedish Medical Birth Register and the time covered were overlapped. Therefore, we included Kallen 2013  for statistical analysis in most outcomes. Kallen 2015 were included for analysis for urogenital malformation, where Kallen 2013 did not provide available data. Sensitivity analysis and subgroup analysis was conducted only for overall congenital malformations for the few studies included in different specific congenital malformations.2 = 82%)  Fig..Fig. 2FoThe summary of meta-analysis of 13 estimates, which analyzed organ-specific congenital malformations, was shown in Fig. The sensitivity analyses revealed no substantial change in the pooled risk estimates upon excluding of any single study Fig. . Meta-reWe performed subgroup analyses regarding exposed period, indication, adjusted for confounders, and risk of bias assessment by ROBINS-I Fig. S–20. The The overall result of this meta-analysis included 18 cohort studies and demonstrated opioid exposure during pregnancy with a 1.3-fold risk of congenital malformations. Additionally, opioid use was associated with increased risks of major malformation, CNS malformation, and limb malformation with an increase of 57, 36, and 127%, respectively. We found no significant relationship between opioid use and cardiovascular malformation, gastrointestinal malformation, ear, face, and neck malformation, respiratory malformation, musculoskeletal malformation, urogenital malformation, orofacial clefts, neural tube defects, gastroschisis, and clubfoot were found. No association in subgroups that opioid was exposed in the first trimester or prescribed for analgesic or antitussive treatment. However, the positive findings were only observed in subgroups that studies provided data unadjusted for confounders or identified critical risk of bias assessment. In subgroups that study provided data adjusted for confounders or identified moderate or serious risk of bias, no association of opioid exposure and overall congenital malformation was found between exposed and not exposed group.Opioids can compound act on the endogenous opioid system, which comprises four G protein-coupled receptors and four major peptide families. They can regulate neuronal function and neurotransmission in human brain, brain stem and other tissues, to effectively prevent the sensation of pain from being transmitted to the brain . PregnanWhen stratified by opioid indications, we found opioid use for analgesic or antitussive treatment did not associate with a higher risk of congenital malformation compared with no exposure. On the contrary, patients who use opioids for abuse or opioid-dependent treatment were more likely to give birth to babies with congenital malformations. The cumulative dose of drug used varied between these indications may explain one of the reasons, which might cause increased blood drug level and risk of congenital malformations. Given that those who were addicted to opioids were at greater risk of misusing prescription opioids and might use more opioids, which is way higher than the therapeutic safe boundary . Only twAfter stratified by whether the study controlled for potential confounders to avoid unpredictable bias introduced by other confounders, we found significant heterogeneity between subgroups. The confounding factors included baseline characteristic such as maternal age, obesity, tobacco use, parity, and so on. Alternatively, the result of confounders adjusted studies showed no association between opioid use and overall congenital malformations, which is different from the pooled result. The potential confounders adjusted in these studies were not consistent. For example, Bateman  and NezvOur findings provide evidence for health professionals to weigh the benefit of opioid along with its potential risks. Also, pregnant women, women intended to get pregnant, or reproductive-aged women at risk of any unintended pregnancy could evaluate the potential risk of opioid during pregnancy. Still, the use of opioid in some situations, especially medication assisted therapy for the treatment of substance use disorder, might provide far greater benefits than risks . Our resOur study has several strengths. To date, this is the first meta-analysis evaluating the association between opioid exposure and the risk of congenital malformations. The meta-analysis included a large sample size of 7,077,709 patients and only cohort studies to reduce recall and selection bias. We also did comprehensive subgroup analyses to evaluate the relationship between opioid use and congenital malformations.Our findings are also subject to several limitations. Firstly, the publication year of the studies included ranged from 1976 to 2021 and might contribute to methodologic bias. However, no significance was observed on meta-regression evaluating the relation of publication year and overall congenital malformations. Secondly, high heterogeneity was detected in most of the outcomes. The reason might be that all included studies are retrospective studies, with the potential for confounding. We performed subgroup analyses to reduce the possible influence. Besides, included studies reported congenital malformations based on several kinds of opioids. Some contained only methadone, and some investigated opioid prescriptions, including hydrocodone, oxycodone, codeine. This might contribute to the heterogeneity of the results. Thirdly, the evaluation of the prediction interval revealed that the current 95%CI produces a positive biased estimate of the overall congenital malformation and major malformation, probably due to the between-study heterogeneity, or to the very low certainty of evidence for the two outcomes. Therefore, large size studies with higher level evidence are needed. Fourthly, most of the studies considered the pregnancies as opioid-exposed by referring to prescriptions during pregnancy, it might be possible that though prescriptions were dispensed while opioids were not taken. Well-designed prospective studies are needed to affirm the findings. Fifthly, few studies reported organ-specific malformations, and the categories of malformations reported were inconsistent. For example, Cleary  reportedIn conclusion, we found that maternal opioid exposure in pregnancy was associated with increased risk of major malformation, CNS malformation, and limb malformation. Opioid exposed in the first trimester or prescribed for analgesic or antitussive treatment did not increase the risk of overall congenital malformation. In studies with moderate or serious risk of bias or studies adjusted for confounders, no association was found between opioid exposure and overall congenital malformation. Therefore, the results should be interpreted in caution.Additional file 1 Table S1. Search terms. Table S2. Summary of risk of bias assessment using ROBINS-I tool. Fig. S1. Eggers’s test of studies examining the association between opioids exposure and the risk of congenital malformations. Fig. S2. Forest plot of association between opioid exposure and major congenital malformation. Fig. S3. Forest plot of association between opioid exposure and central nervous system malformation. Fig. S4. Forest plot of association between opioid exposure and limb malformation. Fig. S5. Forest plot of assoxiation between opioid exposure and cardiovascular malformation. Fig. S6. Forest plot of association between opioid exposure and gastrointestinal malformation. Fig. S7. Forest plot of association between opioid exposure and ear, face, and neck malformation. Fig. S8. Forest plot of association between opioid exposure and respiratory malformation. Fig. S9. Forest plot of association between opioid exposure and musculoskeletal malformation. Fig. S10. Forest plot of association between opioid exposure and urogenital malformation. Fig. S11. Forest plot of association between opioid exposure and orofacial malformation. Fig. S12. Forest plot of association between opioid exposure and neural tube defects. Fig. S13. Forest plot of association between opioid exposure and gastroschisis. Fig. S14. Forest plot of association between opioid exposure and clubfoot. Fig. S15. Meta-regression according to the year of publication. Fig. S16. Forest plot of subgroup analysis of exposed period. Fig. S17. Forest plot of subgroup analysis of indication. Fig. S18: Forest plot of subgroup analysis of adjusted for confounders. Fig. S19. Forest plot of subgroup analysis of risk of bias assessment.


In the first wave of the COVID-19 pandemic, the unknown etiology and treatment of the highly transmissible coronavirus posed considerable threats to public mental health. Many people around the globe turned to religion as an attempt to mitigate their heightened psychological distress, but mixed findings have been obtained regarding the association between the use of religious coping and two psychological symptoms—anxiety and depressive symptoms—widely reported in the initial wave.The present meta-analysis was conducted to resolve the empirical inconsistency by synthesizing this body of studies and identifying both individual and national-level factors that accounted for the inconsistent findings.Following PRISMA guidelines, the literature search and data screening procedures yielded 42 eligible studies, with 25,438 participants  from 24 countries spanning seven world regions.r = 0.2886, p < 0.0001). Although the associations of both general and positive religious coping with psychological symptoms were non-significant , the moderation analysis revealed significant positive associations between positive religious coping and psychological symptoms in two demographic groups who experienced greater pandemic distress than their counterparts: younger participants and female participants.Overall, the results showed that only negative religious coping was positively associated with psychological symptoms (This meta-analysis provides a nuanced understanding of the complex nature of religious coping in the initial wave of the COVID-19 pandemic when the levels of public anxiety and stress were heightened. The exclusive use of religious coping may not be associated with low levels of psychological symptoms, implying the importance of supplementing the deployment of this strategy with an array of other strategies. Therapists of mental health interventions should show their clients how to make good use of positive religious coping together with other strategies, and how to avoid the use of negative religious coping, to handle their psychological problems.https://osf.io/shb32/ Since late 2019, the COVID-19 outbreak has swept swiftly across the globe, affecting more than 650 million people in 228 countries and territories worldwide as of December 2022 , levels During the outbreak of COVID-19, many individuals have resorted to religion for stress relief. Religion constitutes an important part of human civilization and exerts a huge impact on people’s behavior , and relWhen the World Health Organization declared COVID-19 a pandemic in March 2020, the search volume of the keyword “prayer” on Google search engine increased by 30% relative to all other Google searches, soaring to a record-high level . Some scDespite such psychological benefits, there were also studies that yielded null or even contrary findings regarding the association between the use of religious coping and certain mental health criteria. Specifically, some studies have revealed that the frequency of using religious coping was unrelated to stress, anxiety, and depressive symptoms –32, wherTo resolve this body of inconsistent empirical evidence regarding the association between the use of religious coping and psychological symptoms during the first wave of the COVID-19 pandemic, a meta-analysis was conducted to systematically evaluate the findings derived from prior studies and to provide a quantitative estimate of the overall effect size of the religious coping–symptom association , 39. In A review of the literature revealed that three clusters of moderators could account for the between-study differences. First, the dimension of religious coping has been proposed to moderate the magnitude of the coping–symptom association. Earlier research , 41 concTypical examples of PRC are seeking psychological comfort from God or a transcendent force and seeing God as a partner who guides one along the journey of life , 43. FroTypical examples of NRC are perceived conflicts and struggles with God or spirituality, difficulty seeing the meaning of life, doubts about support from God, and anger toward God , 43, 45.Second, the magnitude of the religious coping–symptom association may vary according to age and sex. Our notion stems from studies that have revealed that individuals of different sexes and ages tended to react to stress and to use religious coping in diverse ways. With regard to age differences, older people were found to be more religious , 48. MorWith regard to sex differences, studies have shown that the spouses of male patients with lung cancer tended to use religious coping to handle stress more than those of female patients . SimilarThird, the strength of the religious coping–symptom association was proposed to vary by two religion-related characteristics at the national level: dominant religion and religiosity. COVID-19 has affected many countries with various dominant religions, and thus, the religious beliefs and practices of residents of these countries should differ considerably. According to an extensive global analysis of religion , ChristiIn multinational comparisons, the potential confounding effects of certain national characteristics should be controlled to ensure robust hypothesis testing. Socioeconomic development is a crucial confounding factor. This notion stems from a previous multinational comparison that identified differences in national subjective well-being among countries with distinct levels of socioeconomic development . ResidenA multinational comparison may also be confounded by the between-country differences in the number of confirmed COVID-19 infections and mortality cases . During In summary, the present meta-analysis aimed to quantitatively synthesize the literature investigating religious coping and mental health issues amid the first wave of the COVID-19 pandemic, and to account for potential between-study variations in the associations between religious coping and psychological symptoms by three clusters of moderators .via the Open Science Framework1.The present meta-analysis was conducted to address the research questions regarding the hypothesized variations in the magnitude of the three dimensions of religious coping  and two major indicators of psychological symptoms  experienced during the initial wave of the COVID-19 pandemic. The meta-analysis was performed and reported in compliance with the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) 2020 guidelines . As all a priori. Specifically, inclusion criteria were primary studies that included measures of both religious coping and any of the aforementioned target symptoms. Studies were excluded if at least one of the following criteria were met: (a) no empirical data ; (b) no relevant measures of religious coping or target symptoms; (c) data were collected before the COVID-19 pandemic; (d) duplicate samples; or (e) inadequate information for effect size estimation.To screen potentially relevant eligible articles, eligibility criteria were developed ®), health and medical science , and multi-disciplinary  research fields. Moreover, gray literature  and dissertation databases  were searched for additional primary studies, especially unpublished ones. Finally, scholars were contacted to request unpublished or unavailable works. Requests were made to 24 investigators, of whom 13 provided articles, data, or statistics.Computerized literature searches were first conducted to systematically locate primary studies. These general and discipline-specific databases spanned the social science  were combined using the Boolean operators “AND” and “OR”: (“religious coping” OR “religion coping”) AND (“COVID” OR “coronavirus”) AND . These combined terms were searched in the title, abstract, and subject of potentially relevant works.To obtain as many potentially relevant articles as possible, we did not impose any restrictions regarding the articles’ publication status, publication date, language, study design, or sample characteristics .As the search results were derived from multiple sources, duplicate records were identified and removed using the EndNote software (version 20.4.1). Then, two independent reviewers conducted preliminary relevance screening and perused the full-text articles for further screening and coding.post-hoc meetings.After the pool of eligible studies was finalized, the data extraction task was carried out by the same two reviewers according to a coding manual with full descriptions of the pre-established inclusion and exclusion criteria. The triangulation method was used to enhance inter-reviewer reliability . SpecifiInter-reviewer reliability was evaluated using the concordance statistic of Krippendorff alpha . The resr was chosen as the target effect size for the present meta-analysis because of its high interpretability  were transformed into Pearson r using the esc package written in the R programing language , year of publication, publication status , country in which the study was conducted, dominant religion of the country, gender composition , age composition (mean or median age), measure of religious coping, dimension of religious coping, measure of symptom, type of symptom, and language of the article. Any missing information, clarifications, or verifications were requested from the authors of individual primary studies through email.The religion-related data were extracted from the GALLUP and NationMaster databases , 70. TheMany studies have examined and reported more than one effect size, and it is noteworthy that the problem of dependence could bias the standard errors and statistical inferences of a meta-analysis. This potential methodological problem was addressed using a three-level meta-analysis , taking 2-based homogeneity test was performed to assess heterogeneity across the eligible studies. The presence of between-study heterogeneity was indicated by a significant Cochran’s Q statistic . The level of heterogeneity was indicated by the inconsistency index I2. In line with a widely adopted standard  moderate for an I2 value that fell within the 25–75% range, and (c) high for an I2 value that was greater than 75% .For the major analyses, both the three-level meta-analysis and meta-regression analysis were conducted using RStudio for Windows  with the R package metafor . KrippenThe meta-analysis pooled the effect size estimates from an array of individual studies, and inevitably, the methodological quality varied vastly across the studies. The risk of study bias was evaluated using study quality assessment . The folIt is also important to detect the potential confounding effects of various biases and to verify the robustness of the study conclusions. To detect the possible risk of single-study bias, one-study-removed sensitivity analysis was conducted for all six effect size estimates. This sensitivity analysis examined whether the overall meta-analytic findings would be altered by the omission of any single eligible study. The robustness of the findings would be demonstrated if an overall effect size estimate remained stable across the various stages of study removal. However, significant changes in any of the effect size estimates indicated the presence of single-study bias.Publication bias was estimated using various common statistical techniques. Specifically, potential publication bias was first visualized by a funnel plot, and then Egger’s linear regression test was conducted to identify funnel plot asymmetry . A signip-curve analysis was used to assess the robustness of the present meta-analytic findings by estimating the true effect size  were journal articles, 2% were research reports, and 2% were unpublished theses or dissertations. The pool of eligible studies comprised 43 independent samples with a total of 25,438 participants . In the samples, 58% were females, and the mean age was 36.50 years (SD = 10.67). The samples were drawn from 24 countries spanning seven world regions: 33% from Europe , 26% from the Middle East , 15% from North America , 12% from Asia , 7% from South America (Brazil), 5% from Africa (Egypt and Morocco), and 2% from Oceania .Western religion  was dominant in 47% of the countries, and non-Western religion (Islam and Hinduism) was dominant in 51% of the countries. The remaining 2% were studies that included more than two countries, and thus, the national religion-related data could not be coded. The average national religiosity score of the entire pool of eligible studies was 78.55 (SD = 16.17), and ranged from 32  to 97 (Egypt).ps < 0.0004). In contrast, all the other religious coping–symptom associations were non-significant.The outlier analysis revealed four outliers. Two were identified in the studies that examined GRC, one in the studies that examined PRC, and one in the studies that examined NRC. Using the outlier removal method, the pattern of results was highly similar to that yielded from the full dataset, and the conclusions drawn from both sets of findings were the same (see the right panel of ps < 0.0001), and all the I2 statistics indicated moderate levels of heterogeneity. These results demonstrated that the random-effects model should be used and that moderation analysis should be conducted to identify the sources of heterogeneity.All the Cochran’s Q statistics indicated the presence of heterogeneity , but this association was positive in samples with a larger proportion of younger participants . Moreover, there was a significant positive PRC–symptom association in samples including more female participants , whereas this association was non-significant in samples including fewer female participants .The results of the moderation analysis are summarized in r = 0.1668, SE = 0.0490, p = 0.0007) as well as in countries with a Western dominant religion , but the magnitude of these significant positive associations differed. In addition, the hypothesized positive NRC–symptom association was significant in countries with higher religiosity  but only marginally significant in those with lower religiosity .Two national religion-related characteristics—dominant religion (Western vs. non-Western) and religiosity—were found to moderate the associations between NRC and psychological symptoms. The positive NRC–symptom associations were significant in countries with a non-Western dominant religion (The results indicated that the majority of the eligible studies (91%) had adequate statistical power. Approximately two-thirds of the studies included measures that displayed adequate internal consistency (67%) and included validated measures (65%). However, only 7% of the studies used a random sampling method, and only 2% adopted a longitudinal or prospective design.B = –0.1625, SE = 0.0289, p < 0.0001). Although the GRC–symptom association was non-significant in the majority of studies adopting a cross-sectional design , it is noteworthy that the negative GRC–symptom association was marginally significant in the small subset of studies adopting a longitudinal or prospective design .The results of the moderation analysis revealed that study design had significant moderating effects on the association between GRC and psychological symptoms . The hypothesized positive PRC–symptom association was obtained in studies adopting less reliable measures , but this association was non-significant in studies adopting more reliable measures .Significant moderation effects of scale reliability were also found on the association between PRC and psychological symptoms (ps > 0.05). Similarly, the trim-and-fill techniques did not alter the pooled effect size estimates. Finally, the p-curves of all three religious coping–symptom associations were significant and right-skewed, demonstrating considerable evidence value and the absence of the p-hacking problem. The statistical power of the three religious coping–symptom associations were also high, ranging from 0.96 to 0.99. Taken together, these analyses provided empirical evidence for the robustness of the meta-analytic findings.The one-study-removed sensitivity analysis revealed considerable stability for each of the six effect size estimates, indicating low risk of single-study bias. Regarding publication bias, the results revealed that the Egger’s tests failed to indicate publication bias  and the younger (vs. older) generation generally reporting higher levels of anxiety and depression , 83–86. Previous research has unveiled two major cognitive mechanisms—cognitive reappraisal and perceived coping efficacy—that accounted for the mental health benefits of PRC , 88. TheThe present meta-analysis revealed robust findings regarding the hypothesized positive association between NRC and psychological symptoms. In the literature, three psychological mechanisms have been identified to account for this positive association: life stress, cognitive processes, and delay in seeking professional treatment . Public Apart from demographic differences, national differences in the strength of the NRC–symptom association were also found in both the dominant religion (Western vs. non-Western) and religiosity, especially the latter. The positive NRC–symptom association was weaker in countries characterized by a Western religion and lower in religiosity. This intricate finding may reflect the recent observations in the decline in the level of religiosity in Western countries such as the United States . HoweverThe present findings may have practical implications for mental health professionals. As religious coping is a preferred strategy for many people, facilitators of mental health treatment programs should make good use of PRC and also work to reduce the use of NRC when treating their clients who have experienced mental health issues during the COVID-19 pandemic , 97. As In addition to practical implications, our meta-analytic findings also have implications for researchers. The null findings for the GRC–symptom association suggest that conceptualizing and assessing religious coping as a unidimensional construct should be avoided. Rather, the findings provide further evidence that PRC and NRC are conceptually distinct and thus should be assessed by separate scales, each with a unique set of coping behaviors. In addition, it is important to note that the mental health benefits of GRC were found in a small subset of studies that contained more than a single time point. In this small but significant body of longitudinal or prospective studies, GRC was found to reduce psychological symptoms over time during the COVID-19 pandemic. The overall null findings for the GRC–symptom association may have been yielded by the cross-sectional studies that currently dominate the literature on coping with COVID-19. More longitudinal or prospective studies should be conducted to capture the trajectory of the use of religious coping and an array of other coping strategies to deal with stress as the pandemic unfolds.Finally, our study quality assessment revealed some methodological limitations in the studies included in the present meta-analysis. Very few (7%) of the studies relied on probabilistic sampling for participant recruitment, and even fewer (2%) investigated the use of religious coping over more than one time point; thus, possible changes in strategy deployment and mental health levels over time remain largely unknown. As public anxiety tends to fluctuate across various waves of a disease outbreak , more foThe original contributions presented in this study are included in this article, further inquiries can be directed to the corresponding author.CC contributed to the study design, investigation, supervision and coordination of data coding and all other research tasks, statistical analysis, data interpretation, and wrote the first draft of the manuscript. WY contributed to the investigation, literature search, data interpretation, and wrote portions of the manuscript. Both authors contributed to the article and approved the submitted version.


Growth differentiation factor-15 (GDF-15) and fibroblast growth factor-23 (FGF-23) are considered predictors of the incidence of cardiovascular diseases. The present meta-analysis aimed to elucidate the associations between GDF-15 and FGF-23 in the risk of atrial fibrillation (AF).An electronic search was conducted in the Cochrane Library, PubMed, and Embase databases from inception until February 27, 2021. The study protocol was registered in the PROSPERO database (CRD42020182226).nonlinear = 0.9507), with a RR elevation of 7% for every 20 pg/ml increase in the serum FGF-23 levels (95% CI: 1.02–1.13). No remarkable relationship was found between serum GDF-15 levels and the risk of AF, and the overall RR for the association between a 100 ng/L increment in GDF-15 levels and AF was 1.01 (95% CI: 0.998–1.02).In total, 15 studies that enrolled 36,017 participants were included. Both serum FGF-23 and GDF-15 were elevated in patients with AF. Analysis of categorical variables showed higher serum FGF-23 levels were associated with an increased risk of AF . In contrast, this association was not found with GDF-15 . In dose-response analysis, a linear positive association was noted between serum FGF-23 levels and the risk of AF (P Our study showed a positive linear correlation between serum FGF-23 levels and the risk of AF. However, no significant association was found between GDF-15 and the risk of AF. Further studies are warranted to clarify whether serum FGF-23 levels may be considered in predicting the risk of AF.Systematic Review Registration:http:www.york.ac.uk/inst/crd, identifier CRD42020182226. Atrial fibrillation (AF) is the most prevalent sustained arrhythmia in clinical practice and is associated with high morbidity and mortality , 2. AlthFGF-23 is a bone-derived hormone that plays an essential role in regulating the metabolism of phosphate and 1,25-dihydroxyvitamin D . In addihttp:www.york.ac.uk/inst/crd)-registration number-CRD42020182226.This study was conducted following the guidelines of the Preferred Reporting Item for Systematic Review and Meta-Analysis (PRISMA) . AdditioThe PubMed database, Embase database, and Cochrane database were searched using the following keywords up to February 27, 2021, with no language restriction. The search terms according to PICOS were as follows:Exposure:For GDF-15: “growth differentiation factor 15” OR “macrophage inhibitory cytokine 1” OR “prostate differentiation factor” OR “GDF-15”.For FGF-23: “fibroblast growth factor-23” OR “FGF-23 protein” OR “fibroblast growth factor 23” OR “FGF-23 protein” OR “phosphatonin” OR “tumor-derived hypophosphatemia inducing factor”.Outcomes:For AF: “atrial fibrillation” OR “atrial flutter” OR “atrial arrhythmia” OR “atrial tachycardia”.The detailed description of the search strategy was described in The Endnote X9  database was used to manage all citations. The abstracts of the studies investigating the association between GDF-15 and FGF-23 were reviewed and the full texts were then searched.The inclusion criteria were: (1) The article reported serum GDF-15/FGF-23 levels in the AF and non-AF populations; (2) Studies designed as observational studies  reported the association between baseline serum GDF-15/FGF-23 level and risk of AF, with adjusted odds ratios (OR), relative risk (RR) or hazard ratio (HR), and the corresponding 95% confidence interval (CI), or providing data to calculate these effects size. The exclusion criteria were: (1) articles with incomplete data provided, such as letters, comments, and reviews; (2) the cross-sectional studies were excluded due to the high risk for bias; (3) articles involved specific genetic polymorphisms; (4) AF were expressed at tissue or cell level, such as the degree of structural remodeling.If the same population was used in multiple studies, the most informative article was included.According to the above inclusion criteria, the researchers (Z.Q-T and X-L) independently evaluated the eligibility of the literature. The basic characteristics of each study were extracted, including the first author, year of publication, age, gender, complications, sample size, adjusted estimated effect, 95% confidence interval of each category, and adjustments. The Newcastle-Ottawa Scale (NOS) was used for quality assessment of the articles, with scores ranging from 0 to 9. A higher grade (≥7) indicates a moderate-high quality; otherwise, the articles were regarded as low-quality , 19.2 statistics to estimate the heterogeneity between studies. Low heterogeneity, moderate heterogeneity, and high heterogeneity were defined as I2 <50%, 50–75%, >75%, respectively .The researchers converted the effect measure into its natural logarithm (RR) and calculated the standard error [selog (RR)] according to the corresponding 95% CI. Random-effects models were used considering the potential heterogeneity across studies. GDF-15 and FGF-23 levels were converted into a uniform unit across all included studies . To compare the GDF-15 levels between the AF and control groups, the GDF-15 and FGF-23 levels that were originally expressed as quartiles and medians were converted to mean and standard deviation , 21. Theectively . Review n = 31) are summarized with detailed reasons in A total of 389 publications were initially retrieved . After removing 50 duplicates and 188 irrelevant citations, the full text of the remaining 151 articles were reviewed, and 15 studies (9 for GDF-15 and 6 for FGF-23) were finally included. The flowchart of the study selection is shown in n = 6), two studies were undertaken in China  (n = 2)  , 33.Ascertainment of AF in most studies was mainly conducted through electrocardiography or medical records; Shao's study did not specify the method of AF diagnosis .These studies achieved Newcastle-Ottawa Scale (NOS) scores >6 points, and their estimated quality was acceptable .I2 = 75%] compared to patients without AF, with a significant heterogeneity , and significant heterogeneity was detected  were included in the categorical analysis , 28. Thedetected .I2 = 35%), with no evidence of heterogeneity (P > 0.05)  .I2 = 94%), with substantial evidence of heterogeneity , with no evidence of heterogeneity (P > 0.05)  .non−linear = 0.9507), with an FGF-23 cutoff value of 62 pg/ml indicating a significantly increased risk of AF (I2 = 0%), with no evidence of heterogeneity , the meta-regression was not conducted to assess the source of heterogeneity according to the guidelines  can be found in the article/XL contributed to the study concept, design, and revised the draft. ZT, TS, and SH performed the search strategy, contributed to database research, acquisition of data, and statistical analyses. All authors participated in data analysis, reviewed, and approved the final manuscript.This work was supported in part by the National Natural Science Foundation of China , the Jiangxi Provincial Natural Science Foundation for Youth Scientific Research (No. 20192ACBL21037), and the China Postdoctoral Science Foundation (2021M703724).The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.