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The Cochrane Database of Systematic... May 2023Chronic pain is common in adults, and often has a detrimental impact upon physical ability, well-being, and quality of life. Previous reviews have shown that certain... (Meta-Analysis)
Meta-Analysis Review
BACKGROUND
Chronic pain is common in adults, and often has a detrimental impact upon physical ability, well-being, and quality of life. Previous reviews have shown that certain antidepressants may be effective in reducing pain with some benefit in improving patients' global impression of change for certain chronic pain conditions. However, there has not been a network meta-analysis (NMA) examining all antidepressants across all chronic pain conditions.
OBJECTIVES
To assess the comparative efficacy and safety of antidepressants for adults with chronic pain (except headache).
SEARCH METHODS
We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, AMED and PsycINFO databases, and clinical trials registries, for randomised controlled trials (RCTs) of antidepressants for chronic pain conditions in January 2022.
SELECTION CRITERIA
We included RCTs that examined antidepressants for chronic pain against any comparator. If the comparator was placebo, another medication, another antidepressant, or the same antidepressant at different doses, then we required the study to be double-blind. We included RCTs with active comparators that were unable to be double-blinded (e.g. psychotherapy) but rated them as high risk of bias. We excluded RCTs where the follow-up was less than two weeks and those with fewer than 10 participants in each arm. DATA COLLECTION AND ANALYSIS: Two review authors separately screened, data extracted, and judged risk of bias. We synthesised the data using Bayesian NMA and pairwise meta-analyses for each outcome and ranked the antidepressants in terms of their effectiveness using the surface under the cumulative ranking curve (SUCRA). We primarily used Confidence in Meta-Analysis (CINeMA) and Risk of Bias due to Missing Evidence in Network meta-analysis (ROB-MEN) to assess the certainty of the evidence. Where it was not possible to use CINeMA and ROB-MEN due to the complexity of the networks, we used GRADE to assess the certainty of the evidence. Our primary outcomes were substantial (50%) pain relief, pain intensity, mood, and adverse events. Our secondary outcomes were moderate pain relief (30%), physical function, sleep, quality of life, Patient Global Impression of Change (PGIC), serious adverse events, and withdrawal.
MAIN RESULTS
This review and NMA included 176 studies with a total of 28,664 participants. The majority of studies were placebo-controlled (83), and parallel-armed (141). The most common pain conditions examined were fibromyalgia (59 studies); neuropathic pain (49 studies) and musculoskeletal pain (40 studies). The average length of RCTs was 10 weeks. Seven studies provided no useable data and were omitted from the NMA. The majority of studies measured short-term outcomes only and excluded people with low mood and other mental health conditions. Across efficacy outcomes, duloxetine was consistently the highest-ranked antidepressant with moderate- to high-certainty evidence. In duloxetine studies, standard dose was equally efficacious as high dose for the majority of outcomes. Milnacipran was often ranked as the next most efficacious antidepressant, although the certainty of evidence was lower than that of duloxetine. There was insufficient evidence to draw robust conclusions for the efficacy and safety of any other antidepressant for chronic pain. Primary efficacy outcomes Duloxetine standard dose (60 mg) showed a small to moderate effect for substantial pain relief (odds ratio (OR) 1.91, 95% confidence interval (CI) 1.69 to 2.17; 16 studies, 4490 participants; moderate-certainty evidence) and continuous pain intensity (standardised mean difference (SMD) -0.31, 95% CI -0.39 to -0.24; 18 studies, 4959 participants; moderate-certainty evidence). For pain intensity, milnacipran standard dose (100 mg) also showed a small effect (SMD -0.22, 95% CI -0.39 to 0.06; 4 studies, 1866 participants; moderate-certainty evidence). Mirtazapine (30 mg) had a moderate effect on mood (SMD -0.5, 95% CI -0.78 to -0.22; 1 study, 406 participants; low-certainty evidence), while duloxetine showed a small effect (SMD -0.16, 95% CI -0.22 to -0.1; 26 studies, 7952 participants; moderate-certainty evidence); however it is important to note that most studies excluded participants with mental health conditions, and so average anxiety and depression scores tended to be in the 'normal' or 'subclinical' ranges at baseline already. Secondary efficacy outcomes Across all secondary efficacy outcomes (moderate pain relief, physical function, sleep, quality of life, and PGIC), duloxetine and milnacipran were the highest-ranked antidepressants with moderate-certainty evidence, although effects were small. For both duloxetine and milnacipran, standard doses were as efficacious as high doses. Safety There was very low-certainty evidence for all safety outcomes (adverse events, serious adverse events, and withdrawal) across all antidepressants. We cannot draw any reliable conclusions from the NMAs for these outcomes.
AUTHORS' CONCLUSIONS
Our review and NMAs show that despite studies investigating 25 different antidepressants, the only antidepressant we are certain about for the treatment of chronic pain is duloxetine. Duloxetine was moderately efficacious across all outcomes at standard dose. There is also promising evidence for milnacipran, although further high-quality research is needed to be confident in these conclusions. Evidence for all other antidepressants was low certainty. As RCTs excluded people with low mood, we were unable to establish the effects of antidepressants for people with chronic pain and depression. There is currently no reliable evidence for the long-term efficacy of any antidepressant, and no reliable evidence for the safety of antidepressants for chronic pain at any time point.
Topics: Adult; Humans; Antidepressive Agents; Chronic Pain; Duloxetine Hydrochloride; Milnacipran; Network Meta-Analysis; Pain Management; Randomized Controlled Trials as Topic
PubMed: 37160297
DOI: 10.1002/14651858.CD014682.pub2 -
Journal of Medical Internet Research Feb 2022Nonadherence to medication in tuberculosis (TB) hampers optimal treatment outcomes. Digital health technology (DHT) seems to be a promising approach to managing problems... (Review)
Review
BACKGROUND
Nonadherence to medication in tuberculosis (TB) hampers optimal treatment outcomes. Digital health technology (DHT) seems to be a promising approach to managing problems of nonadherence to medication and improving treatment outcomes.
OBJECTIVE
This paper systematically reviews the effect of DHT in improving medication adherence and treatment outcomes in patients with TB.
METHODS
A literature search in PubMed and Cochrane databases was conducted. Randomized controlled trials (RCTs) that analyzed the effect of DHT interventions on medication adherence outcomes (treatment completion, treatment adherence, missed doses, and noncompleted rate) and treatment outcomes (cure rate and smear conversion) were included. Adult patients with either active or latent TB infection were included. The Jadad score was used for evaluating the study quality. The PRISMA (Preferred Reporting Items for Systematic Reviews and Meta-Analyses) guideline was followed to report study findings.
RESULTS
In all, 16 RCTs were selected from 552 studies found, and 6 types of DHT interventions for TB were identified: 3 RCTs examined video directly observed therapy (VDOT), 1 examined video-observed therapy (VOT), 1 examined an ingestible sensor, 1 examined phone call reminders, 2 examined medication monitor boxes, and 8 examined SMS text message reminders. The outcomes used were treatment adherence, including treatment completion, treatment adherence, missed dose, and noncompleted rate, as well as clinical outcomes, including cure rate and smear conversion. In treatment completion, 4 RCTs (VDOT, VOT, ingestible sensor, SMS reminder) found significant effects, with odds ratios and relative risks (RRs) ranging from 1.10 to 7.69. Treatment adherence was increased in 1 study by SMS reminders (RR 1.05; 95% CI 1.04-1.06), and missed dose was reduced in 1 study by a medication monitor box (mean ratio 0.58; 95% CI 0.42-0.79). In contrast, 3 RCTs of VDOT and 3 RCTs of SMS reminders did not find significant effects for treatment completion. Moreover, no improvement was found in treatment adherence in 1 RCT of VDOT, missed dose in 1 RCT of SMS reminder, and noncompleted rate in 1 RCT of a monitor box, and 2 RCTs of SMS reminders. For clinical outcomes such as cure rate, 2 RCTs reported that phone calls (RR 1.30; 95% CI 1.07-1.59) and SMS reminders (OR 2.47; 95% CI 1.13-5.43) significantly affected cure rates. However, 3 RCTs found that SMS reminders did not have a significant impact on cure rate or smear conversion.
CONCLUSIONS
It was found that DHT interventions can be a promising approach. However, the interventions exhibited variable effects regarding effect direction and the extent of improving TB medication adherence and clinical outcomes. Developing DHT interventions with personalized feedback is required to have a consistent and beneficial effect on medication adherence and outcomes among patients with TB.
Topics: Adult; Biomedical Technology; Cell Phone; Humans; Medication Adherence; Randomized Controlled Trials as Topic; Reminder Systems; Text Messaging; Treatment Outcome; Tuberculosis
PubMed: 35195534
DOI: 10.2196/33062 -
The Cochrane Database of Systematic... Feb 2023Since the previous Cochrane Review on this topic in 2016, debate has continued surrounding a potential role for vitamin D in reducing risk of asthma exacerbation and... (Meta-Analysis)
Meta-Analysis Review
BACKGROUND
Since the previous Cochrane Review on this topic in 2016, debate has continued surrounding a potential role for vitamin D in reducing risk of asthma exacerbation and improving asthma control. We therefore conducted an updated meta-analysis to include data from new trials completed since this date.
OBJECTIVES
To evaluate the effectiveness and safety of administration of vitamin D or its hydroxylated metabolites in reducing the risk of severe asthma exacerbations (defined as those requiring treatment with systemic corticosteroids) and improving asthma symptom control.
SEARCH METHODS
We searched the Cochrane Airways Group Trial Register and reference lists of articles. We contacted the authors of studies in order to identify additional trials. Date of last search: 8 September 2022.
SELECTION CRITERIA
We included double-blind, randomised, placebo-controlled trials of vitamin D in children and adults with asthma evaluating exacerbation risk or asthma symptom control, or both.
DATA COLLECTION AND ANALYSIS
Four review authors independently applied study inclusion criteria, extracted the data, and assessed risk of bias. We obtained missing data from the authors where possible. We reported results with 95% confidence intervals (CIs). The primary outcome was the incidence of severe asthma exacerbations requiring treatment with systemic corticosteroids. Secondary outcomes included the incidence of asthma exacerbations precipitating an emergency department visit or requiring hospital admission, or both, end-study childhood Asthma Control Test (cACT) or Asthma Control Test (ACT) scores, and end-study % predicted forced expiratory volume in one second (FEV1). We performed subgroup analyses to determine whether the effect of vitamin D on risk of asthma exacerbation was modified by baseline vitamin D status, vitamin D dose, frequency of dosing regimen, form of vitamin D given, and age of participants.
MAIN RESULTS
We included 20 studies in this review; 15 trials involving a total of 1155 children and five trials involving a total of 1070 adults contributed data to analyses. Participant ages ranged from 1 to 84 years, with two trials providing data specific to participants under five years (n = 69) and eight trials providing data specific to participants aged 5 to 16 (n = 766). Across the trials, 1245 participants were male and 1229 were female, with two studies not reporting sex distribution. Fifteen trials contributed to the primary outcome analysis of exacerbations requiring systemic corticosteroids. The duration of trials ranged from three to 40 months; all but two investigated effects of administering cholecalciferol (vitamin D3). As in the previous Cochrane Review, the majority of participants had mild to moderate asthma, and profound vitamin D deficiency (25-hydroxyvitamin D (25(OH)D) < 25 nmol/L) at baseline was rare. Administration of vitamin D or its hydroxylated metabolites did not reduce or increase the proportion of participants experiencing one or more asthma exacerbations treated with systemic corticosteroids (odds ratio (OR) 1.04, 95% CI 0.81 to 1.34; I = 0%; 14 studies, 1778 participants; high-quality evidence). This equates to an absolute risk of 226 per 1000 (95% CI 185 to 273) in the pooled vitamin D group, compared to a baseline risk of 219 participants per 1000 in the pooled placebo group. We also found no effect of vitamin D supplementation on the rate of exacerbations requiring systemic corticosteroids (rate ratio 0.86, 95% CI 0.62 to 1.19; I = 60%; 10 studies, 1599 participants; high-quality evidence), or the time to first exacerbation (hazard ratio 0.82, 95% CI 0.59 to 1.15; I = 22%; 3 studies, 850 participants; high-quality evidence). Subgroup analysis did not reveal any evidence of effect modification by baseline vitamin D status, vitamin D dose, frequency of dosing regimen, or age. A single trial investigating administration of calcidiol reported a benefit of the intervention for the primary outcome of asthma control. Vitamin D supplementation did not influence any secondary efficacy outcome meta-analysed, which were all based on moderate- or high-quality evidence. We observed no effect on the incidence of serious adverse events (OR 0.89, 95% CI 0.56 to 1.41; I = 0%; 12 studies, 1556 participants; high-quality evidence). The effect of vitamin D on fatal asthma exacerbations was not estimable, as no such events occurred in any trial. Six studies reported adverse reactions potentially attributable to vitamin D. These occurred across treatment and control arms and included hypercalciuria, hypervitaminosis D, kidney stones, gastrointestinal symptoms and mild itch. In one trial, we could not ascertain the total number of participants with hypercalciuria from the trial report. We assessed three trials as being at high risk of bias in at least one domain; none of these contributed data to the analysis of the outcomes reported above. Sensitivity analyses that excluded these trials from each outcome to which they contributed did not change the null findings.
AUTHORS' CONCLUSIONS
In contrast to findings of our previous Cochrane Review on this topic, this updated review does not find evidence to support a role for vitamin D supplementation or its hydroxylated metabolites to reduce risk of asthma exacerbations or improve asthma control. Participants with severe asthma and those with baseline 25(OH)D concentrations < 25 nmol/L were poorly represented, so further research is warranted here. A single study investigating effects of calcidiol yielded positive results, so further studies investigating effects of this metabolite are needed.
Topics: Adult; Child; Humans; Male; Female; Infant; Child, Preschool; Adolescent; Young Adult; Middle Aged; Aged; Aged, 80 and over; Calcifediol; Hypercalciuria; Disease Progression; Asthma; Vitamins; Adrenal Cortex Hormones; Vitamin D; Cholecalciferol; Anti-Asthmatic Agents; Randomized Controlled Trials as Topic
PubMed: 36744416
DOI: 10.1002/14651858.CD011511.pub3 -
The Cochrane Database of Systematic... Jul 2020Alcohol is consumed by over 2 billion people worldwide. It is a common substance of abuse and its use can lead to more than 200 disorders including hypertension. Alcohol... (Meta-Analysis)
Meta-Analysis
BACKGROUND
Alcohol is consumed by over 2 billion people worldwide. It is a common substance of abuse and its use can lead to more than 200 disorders including hypertension. Alcohol has both acute and chronic effects on blood pressure. This review aimed to quantify the acute effects of different doses of alcohol over time on blood pressure and heart rate in an adult population.
OBJECTIVES
Primary objective To determine short-term dose-related effects of alcohol versus placebo on systolic blood pressure and diastolic blood pressure in healthy and hypertensive adults over 18 years of age. Secondary objective To determine short-term dose-related effects of alcohol versus placebo on heart rate in healthy and hypertensive adults over 18 years of age.
SEARCH METHODS
The Cochrane Hypertension Information Specialist searched the following databases for randomised controlled trials up to March 2019: the Cochrane Hypertension Specialised Register; the Cochrane Central Register of Controlled Trials (CENTRAL; 2019, Issue 2), in the Cochrane Library; MEDLINE (from 1946); Embase (from 1974); the World Health Organization International Clinical Trials Registry Platform; and ClinicalTrials.gov. We also contacted authors of relevant articles regarding further published and unpublished work. These searches had no language restrictions.
SELECTION CRITERIA
Randomised controlled trials (RCTs) comparing effects of a single dose of alcohol versus placebo on blood pressure (BP) or heart rate (HR) in adults (≥ 18 years of age).
DATA COLLECTION AND ANALYSIS
Two review authors (ST and CT) independently extracted data and assessed the quality of included studies. We also contacted trial authors for missing or unclear information. Mean difference (MD) from placebo with 95% confidence interval (CI) was the outcome measure, and a fixed-effect model was used to combine effect sizes across studies. MAIN RESULTS: We included 32 RCTs involving 767 participants. Most of the study participants were male (N = 642) and were healthy. The mean age of participants was 33 years, and mean body weight was 78 kilograms. Low-dose alcohol (< 14 g) within six hours (2 RCTs, N = 28) did not affect BP but did increase HR by 5.1 bpm (95% CI 1.9 to 8.2) (moderate-certainty evidence). Medium-dose alcohol (14 to 28 g) within six hours (10 RCTs, N = 149) decreased systolic blood pressure (SBP) by 5.6 mmHg (95% CI -8.3 to -3.0) and diastolic blood pressure (DBP) by 4.0 mmHg (95% CI -6.0 to -2.0) and increased HR by 4.6 bpm (95% CI 3.1 to 6.1) (moderate-certainty evidence for all). Medium-dose alcohol within 7 to 12 hours (4 RCTs, N = 54) did not affect BP or HR. Medium-dose alcohol > 13 hours after consumption (4 RCTs, N = 66) did not affect BP or HR. High-dose alcohol (> 30 g) within six hours (16 RCTs, N = 418) decreased SBP by 3.5 mmHg (95% CI -6.0 to -1.0), decreased DBP by 1.9 mmHg (95% CI-3.9 to 0.04), and increased HR by 5.8 bpm (95% CI 4.0 to 7.5). The certainty of evidence was moderate for SBP and HR, and was low for DBP. High-dose alcohol within 7 to 12 hours of consumption (3 RCTs, N = 54) decreased SBP by 3.7 mmHg (95% CI -7.0 to -0.5) and DBP by 1.7 mmHg (95% CI -4.6 to 1.8) and increased HR by 6.2 bpm (95% CI 3.0 to 9.3). The certainty of evidence was moderate for SBP and HR, and low for DBP. High-dose alcohol ≥ 13 hours after consumption (4 RCTs, N = 154) increased SBP by 3.7 mmHg (95% CI 2.3 to 5.1), DBP by 2.4 mmHg (95% CI 0.2 to 4.5), and HR by 2.7 bpm (95% CI 0.8 to 4.6) (moderate-certainty evidence for all). AUTHORS' CONCLUSIONS: High-dose alcohol has a biphasic effect on BP; it decreases BP up to 12 hours after consumption and increases BP > 13 hours after consumption. High-dose alcohol increases HR at all times up to 24 hours. Findings of this review are relevant mainly to healthy males, as only small numbers of women were included in the included trials.
Topics: Adult; Aged; Aged, 80 and over; Alcohol Drinking; Alcoholic Beverages; Bias; Blood Pressure; Central Nervous System Depressants; Cross-Over Studies; Ethanol; Female; Heart Rate; Humans; Male; Middle Aged; Randomized Controlled Trials as Topic; Sex Factors; Time Factors; Young Adult
PubMed: 32609894
DOI: 10.1002/14651858.CD012787.pub2 -
European Journal of Preventive... Dec 2023There is good evidence showing that inactivity and walking minimal steps/day increase the risk of cardiovascular (CV) disease and general ill-health. The optimal number... (Meta-Analysis)
Meta-Analysis
AIMS
There is good evidence showing that inactivity and walking minimal steps/day increase the risk of cardiovascular (CV) disease and general ill-health. The optimal number of steps and their role in health is, however, still unclear. Therefore, in this meta-analysis, we aimed to evaluate the relationship between step count and all-cause mortality and CV mortality.
METHODS AND RESULTS
We systematically searched relevant electronic databases from inception until 12 June 2022. The main endpoints were all-cause mortality and CV mortality. An inverse-variance weighted random-effects model was used to calculate the number of steps/day and mortality. Seventeen cohort studies with a total of 226 889 participants (generally healthy or patients at CV risk) with a median follow-up 7.1 years were included in the meta-analysis. A 1000-step increment was associated with a 15% decreased risk of all-cause mortality [hazard ratio (HR) 0.85; 95% confidence interval (CI) 0.81-0.91; P < 0.001], while a 500-step increment was associated with a 7% decrease in CV mortality (HR 0.93; 95% CI 0.91-0.95; P < 0.001). Compared with the reference quartile with median steps/day 3867 (2500-6675), the Quartile 1 (Q1, median steps: 5537), Quartile 2 (Q2, median steps 7370), and Quartile 3 (Q3, median steps 11 529) were associated with lower risk for all-cause mortality (48, 55, and 67%, respectively; P < 0.05, for all). Similarly, compared with the lowest quartile of steps/day used as reference [median steps 2337, interquartile range 1596-4000), higher quartiles of steps/day (Q1 = 3982, Q2 = 6661, and Q3 = 10 413) were linearly associated with a reduced risk of CV mortality (16, 49, and 77%; P < 0.05, for all). Using a restricted cubic splines model, we observed a nonlinear dose-response association between step count and all-cause and CV mortality (Pnonlineraly < 0.001, for both) with a progressively lower risk of mortality with an increased step count.
CONCLUSION
This meta-analysis demonstrates a significant inverse association between daily step count and all-cause mortality and CV mortality with more the better over the cut-off point of 3867 steps/day for all-cause mortality and only 2337 steps for CV mortality.
Topics: Humans; Walking; Cardiovascular Diseases; Cohort Studies; Health Status
PubMed: 37555441
DOI: 10.1093/eurjpc/zwad229 -
The Cochrane Database of Systematic... Feb 2022Description of the condition Malaria, an infectious disease transmitted by the bite of female mosquitoes from several Anopheles species, occurs in 87 countries with... (Meta-Analysis)
Meta-Analysis
BACKGROUND
Description of the condition Malaria, an infectious disease transmitted by the bite of female mosquitoes from several Anopheles species, occurs in 87 countries with ongoing transmission (WHO 2020). The World Health Organization (WHO) estimated that, in 2019, approximately 229 million cases of malaria occurred worldwide, with 94% occurring in the WHO's African region (WHO 2020). Of these malaria cases, an estimated 409,000 deaths occurred globally, with 67% occurring in children under five years of age (WHO 2020). Malaria also negatively impacts the health of women during pregnancy, childbirth, and the postnatal period (WHO 2020). Sulfadoxine/pyrimethamine (SP), an antifolate antimalarial, has been widely used across sub-Saharan Africa as the first-line treatment for uncomplicated malaria since it was first introduced in Malawi in 1993 (Filler 2006). Due to increasing resistance to SP, in 2000 the WHO recommended that one of several artemisinin-based combination therapies (ACTs) be used instead of SP for the treatment of uncomplicated malaria caused by Plasmodium falciparum (Global Partnership to Roll Back Malaria 2001). However, despite these recommendations, SP continues to be advised for intermittent preventive treatment in pregnancy (IPTp) and intermittent preventive treatment in infants (IPTi), whether the person has malaria or not (WHO 2013). Description of the intervention Folate (vitamin B9) includes both naturally occurring folates and folic acid, the fully oxidized monoglutamic form of the vitamin, used in dietary supplements and fortified food. Folate deficiency (e.g. red blood cell (RBC) folate concentrations of less than 305 nanomoles per litre (nmol/L); serum or plasma concentrations of less than 7 nmol/L) is common in many parts of the world and often presents as megaloblastic anaemia, resulting from inadequate intake, increased requirements, reduced absorption, or abnormal metabolism of folate (Bailey 2015; WHO 2015a). Pregnant women have greater folate requirements; inadequate folate intake (evidenced by RBC folate concentrations of less than 400 nanograms per millilitre (ng/mL), or 906 nmol/L) prior to and during the first month of pregnancy increases the risk of neural tube defects, preterm delivery, low birthweight, and fetal growth restriction (Bourassa 2019). The WHO recommends that all women who are trying to conceive consume 400 micrograms (µg) of folic acid daily from the time they begin trying to conceive through to 12 weeks of gestation (WHO 2017). In 2015, the WHO added the dosage of 0.4 mg of folic acid to the essential drug list (WHO 2015c). Alongside daily oral iron (30 mg to 60 mg elemental iron), folic acid supplementation is recommended for pregnant women to prevent neural tube defects, maternal anaemia, puerperal sepsis, low birthweight, and preterm birth in settings where anaemia in pregnant women is a severe public health problem (i.e. where at least 40% of pregnant women have a blood haemoglobin (Hb) concentration of less than 110 g/L). How the intervention might work Potential interactions between folate status and malaria infection The malaria parasite requires folate for survival and growth; this has led to the hypothesis that folate status may influence malaria risk and severity. In rhesus monkeys, folate deficiency has been found to be protective against Plasmodium cynomolgi malaria infection, compared to folate-replete animals (Metz 2007). Alternatively, malaria may induce or exacerbate folate deficiency due to increased folate utilization from haemolysis and fever. Further, folate status measured via RBC folate is not an appropriate biomarker of folate status in malaria-infected individuals since RBC folate values in these individuals are indicative of both the person's stores and the parasite's folate synthesis. A study in Nigeria found that children with malaria infection had significantly higher RBC folate concentrations compared to children without malaria infection, but plasma folate levels were similar (Bradley-Moore 1985). Why it is important to do this review The malaria parasite needs folate for survival and growth in humans. For individuals, adequate folate levels are critical for health and well-being, and for the prevention of anaemia and neural tube defects. Many countries rely on folic acid supplementation to ensure adequate folate status in at-risk populations. Different formulations for folic acid supplements are available in many international settings, with dosages ranging from 400 µg to 5 mg. Evaluating folic acid dosage levels used in supplementation efforts may increase public health understanding of its potential impacts on malaria risk and severity and on treatment failures. Examining folic acid interactions with antifolate antimalarial medications and with malaria disease progression may help countries in malaria-endemic areas determine what are the most appropriate lower dose folic acid formulations for at-risk populations. The WHO has highlighted the limited evidence available and has indicated the need for further research on biomarkers of folate status, particularly interactions between RBC folate concentrations and tuberculosis, human immunodeficiency virus (HIV), and antifolate antimalarial drugs (WHO 2015b). An earlier Cochrane Review assessed the effects and safety of iron supplementation, with or without folic acid, in children living in hyperendemic or holoendemic malaria areas; it demonstrated that iron supplementation did not increase the risk of malaria, as indicated by fever and the presence of parasites in the blood (Neuberger 2016). Further, this review stated that folic acid may interfere with the efficacy of SP; however, the efficacy and safety of folic acid supplementation on these outcomes has not been established. This review will provide evidence on the effectiveness of daily folic acid supplementation in healthy and malaria-infected individuals living in malaria-endemic areas. Additionally, it will contribute to achieving both the WHO Global Technical Strategy for Malaria 2016-2030 (WHO 2015d), and United Nations Sustainable Development Goal 3 (to ensure healthy lives and to promote well-being for all of all ages) (United Nations 2021), and evaluating whether the potential effects of folic acid supplementation, at different doses (e.g. 0.4 mg, 1 mg, 5 mg daily), interferes with the effect of drugs used for prevention or treatment of malaria.
OBJECTIVES
To examine the effects of folic acid supplementation, at various doses, on malaria susceptibility (risk of infection) and severity among people living in areas with various degrees of malaria endemicity. We will examine the interaction between folic acid supplements and antifolate antimalarial drugs. Specifically, we will aim to answer the following. Among uninfected people living in malaria endemic areas, who are taking or not taking antifolate antimalarials for malaria prophylaxis, does taking a folic acid-containing supplement increase susceptibility to or severity of malaria infection? Among people with malaria infection who are being treated with antifolate antimalarials, does folic acid supplementation increase the risk of treatment failure?
METHODS
Criteria for considering studies for this review Types of studies Inclusion criteria Randomized controlled trials (RCTs) Quasi-RCTs with randomization at the individual or cluster level conducted in malaria-endemic areas (areas with ongoing, local malaria transmission, including areas approaching elimination, as listed in the World Malaria Report 2020) (WHO 2020) Exclusion criteria Ecological studies Observational studies In vivo/in vitro studies Economic studies Systematic literature reviews and meta-analyses (relevant systematic literature reviews and meta-analyses will be excluded but flagged for grey literature screening) Types of participants Inclusion criteria Individuals of any age or gender, living in a malaria endemic area, who are taking antifolate antimalarial medications (including but not limited to sulfadoxine/pyrimethamine (SP), pyrimethamine-dapsone, pyrimethamine, chloroquine and proguanil, cotrimoxazole) for the prevention or treatment of malaria (studies will be included if more than 70% of the participants live in malaria-endemic regions) Studies assessing participants with or without anaemia and with or without malaria parasitaemia at baseline will be included Exclusion criteria Individuals not taking antifolate antimalarial medications for prevention or treatment of malaria Individuals living in non-malaria endemic areas Types of interventions Inclusion criteria Folic acid supplementation Form: in tablet, capsule, dispersible tablet at any dose, during administration, or periodically Timing: during, before, or after (within a period of four to six weeks) administration of antifolate antimalarials Iron-folic acid supplementation Folic acid supplementation in combination with co-interventions that are identical between the intervention and control groups. Co-interventions include: anthelminthic treatment; multivitamin or multiple micronutrient supplementation; 5-methyltetrahydrofolate supplementation. Exclusion criteria Folate through folate-fortified water Folic acid administered through large-scale fortification of rice, wheat, or maize Comparators Placebo No treatment No folic acid/different doses of folic acid Iron Types of outcome measures Primary outcomes Uncomplicated malaria (defined as a history of fever with parasitological confirmation; acceptable parasitological confirmation will include rapid diagnostic tests (RDTs), malaria smears, or nucleic acid detection (i.e. polymerase chain reaction (PCR), loop-mediated isothermal amplification (LAMP), etc.)) (WHO 2010). This outcome is relevant for patients without malaria, given antifolate antimalarials for malaria prophylaxis. Severe malaria (defined as any case with cerebral malaria or acute P. falciparum malaria, with signs of severity or evidence of vital organ dysfunction, or both) (WHO 2010). This outcome is relevant for patients without malaria, given antifolate antimalarials for malaria prophylaxis. Parasite clearance (any Plasmodium species), defined as the time it takes for a patient who tests positive at enrolment and is treated to become smear-negative or PCR negative. This outcome is relevant for patients with malaria, treated with antifolate antimalarials. Treatment failure (defined as the inability to clear malaria parasitaemia or prevent recrudescence after administration of antimalarial medicine, regardless of whether clinical symptoms are resolved) (WHO 2019). This outcome is relevant for patients with malaria, treated with antifolate antimalarials. Secondary outcomes Duration of parasitaemia Parasite density Haemoglobin (Hb) concentrations (g/L) Anaemia: severe anaemia (defined as Hb less than 70 g/L in pregnant women and children aged six to 59 months; and Hb less than 80 g/L in other populations); moderate anaemia (defined as Hb less than 100 g/L in pregnant women and children aged six to 59 months; and less than 110 g/L in others) Death from any cause Among pregnant women: stillbirth (at less than 28 weeks gestation); low birthweight (less than 2500 g); active placental malaria (defined as Plasmodium detected in placental blood by smear or PCR, or by Plasmodium detected on impression smear or placental histology). Search methods for identification of studies A search will be conducted to identify completed and ongoing studies, without date or language restrictions. Electronic searches A search strategy will be designed to include the appropriate subject headings and text word terms related to each intervention of interest and study design of interest (see Appendix 1). Searches will be broken down by these two criteria (intervention of interest and study design of interest) to allow for ease of prioritization, if necessary. The study design filters recommended by the Scottish Intercollegiate Guidelines Network (SIGN), and those designed by Cochrane for identifying clinical trials for MEDLINE and Embase, will be used (SIGN 2020). There will be no date or language restrictions. Non-English articles identified for inclusion will be translated into English. If translations are not possible, advice will be requested from the Cochrane Infectious Diseases Group and the record will be stored in the "Awaiting assessment" section of the review until a translation is available. The following electronic databases will be searched for primary studies. Cochrane Central Register of Controlled Trials. Cumulative Index to Nursing and Allied Health Literature (CINAHL). Embase. MEDLINE. Scopus. Web of Science (both the Social Science Citation Index and the Science Citation Index). We will conduct manual searches of ClinicalTrials.gov, the International Clinical Trials Registry Platform (ICTRP), and the United Nations Children's Fund (UNICEF) Evaluation and Research Database (ERD), in order to identify relevant ongoing or planned trials, abstracts, and full-text reports of evaluations, studies, and surveys related to programmes on folic acid supplementation in malaria-endemic areas. Additionally, manual searches of grey literature to identify RCTs that have not yet been published but are potentially eligible for inclusion will be conducted in the following sources. Global Index Medicus (GIM). African Index Medicus (AIM). Index Medicus for the Eastern Mediterranean Region (IMEMR). Latin American & Caribbean Health Sciences Literature (LILACS). Pan American Health Organization (PAHO). Western Pacific Region Index Medicus (WPRO). Index Medicus for the South-East Asian Region (IMSEAR). The Spanish Bibliographic Index in Health Sciences (IBECS) (ibecs.isciii.es/). Indian Journal of Medical Research (IJMR) (journals.lww.com/ijmr/pages/default.aspx). Native Health Database (nativehealthdatabase.net/). Scielo (www.scielo.br/). Searching other resources Handsearches of the five journals with the highest number of included studies in the last 12 months will be conducted to capture any relevant articles that may not have been indexed in the databases at the time of the search. We will contact the authors of included studies and will check reference lists of included papers for the identification of additional records. For assistance in identifying ongoing or unpublished studies, we will contact the Division of Nutrition, Physical Activity, and Obesity (DNPAO) and the Division of Parasitic Diseases and Malaria (DPDM) of the CDC, the United Nations World Food Programme (WFP), Nutrition International (NI), Global Alliance for Improved Nutrition (GAIN), and Hellen Keller International (HKI). Data collection and analysis Selection of studies Two review authors will independently screen the titles and abstracts of articles retrieved by each search to assess eligibility, as determined by the inclusion and exclusion criteria. Studies deemed eligible for inclusion by both review authors in the abstract screening phase will advance to the full-text screening phase, and full-text copies of all eligible papers will be retrieved. If full articles cannot be obtained, we will attempt to contact the authors to obtain further details of the studies. If such information is not obtained, we will classify the study as "awaiting assessment" until further information is published or made available to us. The same two review authors will independently assess the eligibility of full-text articles for inclusion in the systematic review. If any discrepancies occur between the studies selected by the two review authors, a third review author will provide arbitration. Each trial will be scrutinized to identify multiple publications from the same data set, and the justification for excluded trials will be documented. A PRISMA flow diagram of the study selection process will be presented to provide information on the number of records identified in the literature searches, the number of studies included and excluded, and the reasons for exclusion (Moher 2009). The list of excluded studies, along with their reasons for exclusion at the full-text screening phase, will also be created. Data extraction and management Two review authors will independently extract data for the final list of included studies using a standardized data specification form. Discrepancies observed between the data extracted by the two authors will be resolved by involving a third review author and reaching a consensus. Information will be extracted on study design components, baseline participant characteristics, intervention characteristics, and outcomes. For individually randomized trials, we will record the number of participants experiencing the event and the number analyzed in each treatment group or the effect estimate reported (e.g. risk ratio (RR)) for dichotomous outcome measures. For count data, we will record the number of events and the number of person-months of follow-up in each group. If the number of person-months is not reported, the product of the duration of follow-up and the number of children evaluated will be used to estimate this figure. We will calculate the rate ratio and standard error (SE) for each study. Zero events will be replaced by 0.5. We will extract both adjusted and unadjusted covariate incidence rate ratios if they are reported in the original studies. For continuous data, we will extract means (arithmetic or geometric) and a measure of variance (standard deviation (SD), SE, or confidence interval (CI)), percentage or mean change from baseline, and the numbers analyzed in each group. SDs will be computed from SEs or 95% CIs, assuming a normal distribution of the values. Haemoglobin values in g/dL will be calculated by multiplying haematocrit or packed cell volume values by 0.34, and studies reporting haemoglobin values in g/dL will be converted to g/L. In cluster-randomized trials, we will record the unit of randomization (e.g. household, compound, sector, or village), the number of clusters in the trial, and the average cluster size. The statistical methods used to analyze the trials will be documented, along with details describing whether these methods adjusted for clustering or other covariates. We plan to extract estimates of the intra-cluster correlation coefficient (ICC) for each outcome. Where results are adjusted for clustering, we will extract the treatment effect estimate and the SD or CI. If the results are not adjusted for clustering, we will extract the data reported. Assessment of risk of bias in included studies Two review authors (KSC, LFY) will independently assess the risk of bias for each included trial using the Cochrane 'Risk of bias 2' tool (RoB 2) for randomized studies (Sterne 2019). Judgements about the risk of bias of included studies will be made according to the recommendations outlined in the Cochrane Handbook for Systematic Reviews of Interventions (Higgins 2021). Disagreements will be resolved by discussion, or by involving a third review author. The interest of our review will be to assess the effect of assignment to the interventions at baseline. We will evaluate each primary outcome using the RoB2 tool. The five domains of the Cochrane RoB2 tool include the following. Bias arising from the randomization process. Bias due to deviations from intended interventions. Bias due to missing outcome data. Bias in measurement of the outcome. Bias in selection of the reported result. Each domain of the RoB2 tool comprises the following. A series of 'signalling' questions. A judgement about the risk of bias for the domain, facilitated by an algorithm that maps responses to the signalling questions to a proposed judgement. Free-text boxes to justify responses to the signalling questions and 'Risk of bias' judgements. An option to predict (and explain) the likely direction of bias. Responses to signalling questions elicit information relevant to an assessment of the risk of bias. These response options are as follows. Yes (may indicate either low or high risk of bias, depending on the most natural way to ask the question). Probably yes. Probably no. No. No information (may indicate no evidence of that problem or an absence of information leading to concerns about there being a problem). Based on the answer to the signalling question, a 'Risk of bias' judgement is assigned to each domain. These judgements include one of the following. High risk of bias Low risk of bias Some concerns To generate the risk of bias judgement for each domain in the randomized studies, we will use the Excel template, available at www.riskofbias.info/welcome/rob-2-0-tool/current-version-of-rob-2. This file will be stored on a scientific data website, available to readers. Risk of bias in cluster randomized controlled trials For the cluster randomized trials, we will be using the RoB2 tool to analyze the five standard domains listed above along with Domain 1b (bias arising from the timing of identification or recruitment of participants) and its related signalling questions. To generate the risk of bias judgement for each domain in the cluster RCTs, we will use the Excel template available at https://sites.google.com/site/riskofbiastool/welcome/rob-2-0-tool/rob-2-for-cluster-randomized-trials. This file will be stored on a scientific data website, available to readers. Risk of bias in cross-over randomized controlled trials For cross-over randomized trials, we will be using the RoB2 tool to analyze the five standard domains listed above along with Domain 2 (bias due to deviations from intended interventions), and Domain 3 (bias due to missing outcome data), and their respective signalling questions. To generate the risk of bias judgement for each domain in the cross-over RCTs, we will use the Excel template, available at https://sites.google.com/site/riskofbiastool/welcome/rob-2-0-tool/rob-2-for-crossover-trials, for each risk of bias judgement of cross-over randomized studies. This file will be stored on a scientific data website, available to readers. Overall risk of bias The overall 'Risk of bias' judgement for each specific trial being assessed will be based on each domain-level judgement. The overall judgements include the following. Low risk of bias (the trial is judged to be at low risk of bias for all domains). Some concerns (the trial is judged to raise some concerns in at least one domain but is not judged to be at high risk of bias for any domain). High risk of bias (the trial is judged to be at high risk of bias in at least one domain, or is judged to have some concerns for multiple domains in a way that substantially lowers confidence in the result). The 'risk of bias' assessments will inform our GRADE evaluations of the certainty of evidence for our primary outcomes presented in the 'Summary of findings' tables and will also be used to inform the sensitivity analyses; (see Sensitivity analysis). If there is insufficient information in study reports to enable an assessment of the risk of bias, studies will be classified as "awaiting assessment" until further information is published or made available to us. Measures of treatment effect Dichotomous data For dichotomous data, we will present proportions and, for two-group comparisons, results as average RR or odds ratio (OR) with 95% CIs. Ordered categorical data Continuous data We will report results for continuous outcomes as the mean difference (MD) with 95% CIs, if outcomes are measured in the same way between trials. Where some studies have reported endpoint data and others have reported change-from-baseline data (with errors), we will combine these in the meta-analysis, if the outcomes were reported using the same scale. We will use the standardized mean difference (SMD), with 95% CIs, to combine trials that measured the same outcome but used different methods. If we do not find three or more studies for a pooled analysis, we will summarize the results in a narrative form. Unit of analysis issues Cluster-randomized trials We plan to combine results from both cluster-randomized and individually randomized studies, providing there is little heterogeneity between the studies. If the authors of cluster-randomized trials conducted their analyses at a different level from that of allocation, and they have not appropriately accounted for the cluster design in their analyses, we will calculate the trials' effective sample sizes to account for the effect of clustering in data. When one or more cluster-RCT reports RRs adjusted for clustering, we will compute cluster-adjusted SEs for the other trials. When none of the cluster-RCTs provide cluster-adjusted RRs, we will adjust the sample size for clustering. We will divide, by the estimated design effects (DE), the number of events and number evaluated for dichotomous outcomes and the number evaluated for continuous outcomes, where DE = 1 + ((average cluster size 1) * ICC). The derivation of the estimated ICCs and DEs will be reported. We will utilize the intra-cluster correlation coefficient (ICC), derived from the trial (if available), or from another source (e.g., using the ICCs derived from other, similar trials) and then calculate the design effect with the formula provided in the Cochrane Handbook for Systematic Reviews of Interventions (Higgins 2021). If this approach is used, we will report it and undertake sensitivity analysis to investigate the effect of variations in ICC. Studies with more than two treatment groups If we identify studies with more than two intervention groups (multi-arm studies), where possible we will combine groups to create a single pair-wise comparison or use the methods set out in the Cochrane Handbook to avoid double counting study participants (Higgins 2021). For the subgroup analyses, when the control group was shared by two or more study arms, we will divide the control group (events and total population) over the number of relevant subgroups to avoid double counting the participants. Trials with several study arms can be included more than once for different comparisons. Cross-over trials From cross-over trials, we will consider the first period of measurement only and will analyze the results together with parallel-group studies. Multiple outcome events In several outcomes, a participant might experience more than one outcome event during the trial period. For all outcomes, we will extract the number of participants with at least one event. Dealing with missing data We will contact the trial authors if the available data are unclear, missing, or reported in a format that is different from the format needed. We aim to perform a 'per protocol' or 'as observed' analysis; otherwise, we will perform a complete case analysis. This means that for treatment failure, we will base the analyses on the participants who received treatment and the number of participants for which there was an inability to clear malarial parasitaemia or prevent recrudescence after administration of an antimalarial medicine reported in the studies. Assessment of heterogeneity Heterogeneity in the results of the trials will be assessed by visually examining the forest plot to detect non-overlapping CIs, using the Chi2 test of heterogeneity (where a P value of less than 0.1 indicates statistical significance) and the I2 statistic of inconsistency (with a value of greater than 50% denoting moderate levels of heterogeneity). When statistical heterogeneity is present, we will investigate the reasons for it, using subgroup analysis. Assessment of reporting biases We will construct a funnel plot to assess the effect of small studies for the main outcome (when including more than 10 trials). Data synthesis The primary analysis will include all eligible studies that provide data regardless of the overall risk of bias as assessed by the RoB2 tool. Analyses will be conducted using Review Manager 5.4 (Review Manager 2020). Cluster-RCTs will be included in the main analysis after adjustment for clustering (see the previous section on cluster-RCTs). The meta-analysis will be performed using the Mantel-Haenszel random-effects model or the generic inverse variance method (when adjustment for clustering is performed by adjusting SEs), as appropriate. Subgroup analysis and investigation of heterogeneity The overall risk of bias will not be used as the basis in conducting our subgroup analyses. However, where data are available, we plan to conduct the following subgroup analyses, independent of heterogeneity. Dose of folic acid supplementation: higher doses (4 mg or more, daily) versus lower doses (less than 4 mg, daily). Moderate-severe anaemia at baseline (mean haemoglobin of participants in a trial at baseline below 100 g/L for pregnant women and children aged six to 59 months, and below 110 g/L for other populations) versus normal at baseline (mean haemoglobin above 100 g/L for pregnant women and children aged six to 59 months, and above 110 g/L for other populations). Antimalarial drug resistance to parasite: known resistance versus no resistance versus unknown/mixed/unreported parasite resistance. Folate status at baseline: Deficient (e.g. RBC folate concentration of less than 305 nmol/L, or serum folate concentration of less than 7nmol/L) and Insufficient (e.g. RBC folate concentration from 305 to less than 906 nmol/L, or serum folate concentration from 7 to less than 25 nmol/L) versus Sufficient (e.g. RBC folate concentration above 906 nmol/L, or serum folate concentration above 25 nmol/L). Presence of anaemia at baseline: yes versus no. Mandatory fortification status: yes, versus no (voluntary or none). We will only use the primary outcomes in any subgroup analyses, and we will limit subgroup analyses to those outcomes for which three or more trials contributed data. Comparisons between subgroups will be performed using Review Manager 5.4 (Review Manager 2020). Sensitivity analysis We will perform a sensitivity analysis, using the risk of bias as a variable to explore the robustness of the findings in our primary outcomes. We will verify the behaviour of our estimators by adding and removing studies with a high risk of bias overall from the analysis. That is, studies with a low risk of bias versus studies with a high risk of bias. Summary of findings and assessment of the certainty of the evidence For the assessment across studies, we will use the GRADE approach, as outlined in (Schünemann 2021). We will use the five GRADE considerations (study limitations based on RoB2 judgements, consistency of effect, imprecision, indirectness, and publication bias) to assess the certainty of the body of evidence as it relates to the studies which contribute data to the meta-analyses for the primary outcomes. The GRADEpro Guideline Development Tool (GRADEpro) will be used to import data from Review Manager 5.4 (Review Manager 2020) to create 'Summary of Findings' tables. The primary outcomes for the main comparison will be listed with estimates of relative effects, along with the number of participants and studies contributing data for those outcomes. These tables will provide outcome-specific information concerning the overall certainty of evidence from studies included in the comparison, the magnitude of the effect of the interventions examined, and the sum of available data on the outcomes we considered. We will include only primary outcomes in the summary of findings tables. For each individual outcome, two review authors (KSC, LFY) will independently assess the certainty of the evidence using the GRADE approach (Balshem 2011). For assessments of the overall certainty of evidence for each outcome that includes pooled data from included trials, we will downgrade the evidence from 'high certainty' by one level for serious (or by two for very serious) study limitations (risk of bias, indirectness of evidence, serious inconsistency, imprecision of effect estimates, or potential publication bias).
Topics: Child; Infant; Pregnancy; Infant, Newborn; Female; Humans; Child, Preschool; Antimalarials; Sulfadoxine; Pyrimethamine; Folic Acid Antagonists; Birth Weight; Parasitemia; Vitamins; Folic Acid; Anemia; Neural Tube Defects; Dietary Supplements; Iron; Recurrence
PubMed: 36321557
DOI: 10.1002/14651858.CD014217 -
Frontiers in Psychiatry 2023Psychedelic-assisted therapy [e.g., with lysergic acid diethylamide (LSD)] has shown promising results as treatment for substance use disorders (SUDs). Previous...
BACKGROUND
Psychedelic-assisted therapy [e.g., with lysergic acid diethylamide (LSD)] has shown promising results as treatment for substance use disorders (SUDs). Previous systematic reviews assessing the efficacy of psilocybin in SUDs only included clinical trials conducted in the last 25 years, but they may have missed clinical trials assessing the efficacy of psilocybin that were conducted before the 1980s, given much research has been done with psychedelics in the mid-20th century. In this systematic review, we specifically assessed the efficacy of psilocybin in patients with a SUD or non-substance-related disorder with no publication date restrictions in our search strategy.
METHODS
A systematic literature search was performed according to Preferred Reporting Items for Systematic reviews and Meta-Analysis (PRISMA) guidelines from the earliest published manuscript up to September 2, 2022, in seven electronic databases, including clinical trials in patients with a SUD or non-substance-related disorder evaluating the efficacy of psilocybin.
RESULTS
A total of four studies (six articles, of which two articles were long-term follow-up results from the same trial) were included in this systematic review. Psilocybin-assisted therapy was administered to = 151 patients in a dose ranging from 6 to 40 mg. Three studies focused on alcohol use disorder, and one study on tobacco use disorder. In a pilot study ( = 10), the percentage of heavy drinking days decreased significantly between baseline and weeks 5-12 (mean difference of 26.0, 95% CI = 8.7-43.2, = 0.008). In another single-arm study ( = 31), 32% (10/31) became completely abstinent from alcohol (mean duration of follow-up 6 years). In a double-blind, placebo-controlled randomized controlled trial (RCT, = 95), the percentage of heavy drinking days during the 32-week double-blind period was significantly lower for psilocybin compared to placebo (mean difference of 13.9, 95% CI = 3.0-24.7, = 0.01). In a pilot study ( = 15), the 7-day point prevalence of smoking abstinence at 26 weeks was 80% (12/15), and at 52 weeks 67% (10/15).
CONCLUSION
Only one RCT and three small clinical trials were identified assessing the efficacy of psilocybin combined with some form of psychotherapy in patients with alcohol and tobacco use disorder. All four clinical trials indicated a beneficial effect of psilocybin-assisted therapy on SUD symptoms. Larger RCTs in patients with SUDs need to evaluate whether psilocybin-assisted therapy is effective in patients with SUD.
PubMed: 36846225
DOI: 10.3389/fpsyt.2023.1134454 -
Clinical Toxicology (Philadelphia, Pa.) Dec 2021The use of activated charcoal in poisoning remains both a pillar of modern toxicology and a source of debate. Following the publication of the joint position statements...
INTRODUCTION
The use of activated charcoal in poisoning remains both a pillar of modern toxicology and a source of debate. Following the publication of the joint position statements on the use of single-dose and multiple-dose activated charcoal by the American Academy of Clinical Toxicology and the European Association of Poison Centres and Clinical Toxicologists, the routine use of activated charcoal declined. Over subsequent years, many new pharmaceuticals became available in modified or alternative-release formulations and additional data on gastric emptying time in poisoning was published, challenging previous assumptions about absorption kinetics. The American Academy of Clinical Toxicology, the European Association of Poison Centres and Clinical Toxicologists and the Asia Pacific Association of Medical Toxicology founded the Clinical Toxicology Recommendations Collaborative to create a framework for evidence-based recommendations for the management of poisoned patients. The activated charcoal workgroup of the Clinical Toxicology Recommendations Collaborative was tasked with reviewing systematically the evidence pertaining to the use of activated charcoal in poisoning in order to update the previous recommendations.
OBJECTIVES
The main objective was: Does oral activated charcoal given to adults or children prevent toxicity or improve clinical outcome and survival of poisoned patients compared to those who do not receive charcoal? Secondary objectives were to evaluate pharmacokinetic outcomes, the role of cathartics, and adverse events to charcoal administration. This systematic review summarizes the available evidence on the efficacy of activated charcoal.
METHODS
A medical librarian created a systematic search strategy for Medline (Ovid), subsequently translated for Embase ( Ovid), CINAHL ( EBSCO), BIOSIS Previews ( Ovid), Web of Science, Scopus, and the Cochrane Library/DARE. All databases were searched from inception to December 31, 2019. There were no language limitations. One author screened all citations identified in the search based on predefined inclusion/exclusion criteria. Excluded citations were confirmed by an additional author and remaining articles were obtained in full text and evaluated by at least two authors for inclusion. All authors cross-referenced full-text articles to identify articles missed in the searches. Data from included articles were extracted by the authors on a standardized spreadsheet and two authors used the GRADE methodology to independently assess the quality and risk of bias of each included study.
RESULTS
From 22,950 titles originally identified, the final data set consisted of 296 human studies, 118 animal studies, and 145 studies. Also included were 71 human and two animal studies that reported adverse events. The quality was judged to have a Low or Very Low GRADE in 469 (83%) of the studies. Ninety studies were judged to be of Moderate or High GRADE. The higher GRADE studies reported on the following drugs: paracetamol (acetaminophen), phenobarbital, carbamazepine, cardiac glycosides (digoxin and oleander), ethanol, iron, salicylates, theophylline, tricyclic antidepressants, and valproate. Data on newer pharmaceuticals not reviewed in the previous American Academy of Clinical Toxicology/European Association of Poison Centres and Clinical Toxicologists statements such as quetiapine, olanzapine, citalopram, and Factor Xa inhibitors were included. No studies on the optimal dosing for either single-dose or multiple-dose activated charcoal were found. In the reviewed clinical data, the time of administration of the first dose of charcoal was beyond one hour in 97% ( = 1006 individuals), beyond two hours in 36% ( = 491 individuals), and beyond 12 h in 4% ( = 43 individuals) whereas the timing of the first dose in controlled studies was within one hour of ingestion in 48% ( = 2359 individuals) and beyond two hours in 36% ( = 484) of individuals.
CONCLUSIONS
This systematic review found heterogenous data. The higher GRADE data was focused on a few select poisonings, while studies that addressed patients with unknown and or mixed ingestions were hampered by low rates of clinically meaningful toxicity or death. Despite these limitations, they reported a benefit of activated charcoal beyond one hour in many clinical scenarios.
Topics: Acetaminophen; Animals; Carbamazepine; Charcoal; Decontamination; Drug Overdose; Humans
PubMed: 34424785
DOI: 10.1080/15563650.2021.1961144 -
The Cochrane Database of Systematic... Jan 2020Increased intracranial pressure has been shown to be strongly associated with poor neurological outcomes and mortality for patients with acute traumatic brain injury.... (Meta-Analysis)
Meta-Analysis
BACKGROUND
Increased intracranial pressure has been shown to be strongly associated with poor neurological outcomes and mortality for patients with acute traumatic brain injury. Currently, most efforts to treat these injuries focus on controlling the intracranial pressure. Hypertonic saline is a hyperosmolar therapy that is used in traumatic brain injury to reduce intracranial pressure. The effectiveness of hypertonic saline compared with other intracranial pressure-lowering agents in the management of acute traumatic brain injury is still debated, both in the short and the long term.
OBJECTIVES
To assess the comparative efficacy and safety of hypertonic saline versus other intracranial pressure-lowering agents in the management of acute traumatic brain injury.
SEARCH METHODS
We searched Cochrane Injuries' Specialised Register, CENTRAL, PubMed, Embase Classic+Embase, ISI Web of Science: Science Citation Index and Conference Proceedings Citation Index-Science, as well as trials registers, on 11 December 2019. We supplemented these searches with searches of four major Chinese databases on 19 September 2018. We also checked bibliographies, and contacted trial authors to identify additional trials.
SELECTION CRITERIA
We sought to identify all randomised controlled trials (RCTs) of hypertonic saline versus other intracranial pressure-lowering agents for people with acute traumatic brain injury of any severity. We excluded cross-over trials as incompatible with assessing long-term outcomes.
DATA COLLECTION AND ANALYSIS
Two review authors independently screened search results to identify potentially eligible trials and extracted data using a standard data extraction form. Outcome measures included: mortality at end of follow-up (all-cause); death or disability (as measured by the Glasgow Outcome Scale (GOS)); uncontrolled intracranial pressure (defined as failure to decrease the intracranial pressure to target and/or requiring additional intervention); and adverse events e.g. rebound phenomena; pulmonary oedema; acute renal failure during treatment).
MAIN RESULTS
Six trials, involving data from 287 people, met the inclusion criteria. The majority of participants (91%) had a diagnosis of severe traumatic brain injury. We had concerns about particular domains of risk of bias in each trial, as physicians were not reliably blinded to allocation, two trials contained participants with conditions other than traumatic brain injury and in one trial, we had concerns about missing data for important outcomes. The original protocol was available for only one trial and other trials (where registered) were registered retrospectively. Meta-analysis for both the primary outcome (mortality at final follow-up) and for 'poor outcome' as per conventionally dichotomised GOS criteria, was only possible for two trials. Synthesis of long-term outcomes was inhibited by the fact that two trials ceased data collection within two hours of a single bolus dose of an intracranial pressure-lowering agent and one at discharge from the intensive care unit (ICU). Only three trials collected data after participants were released from hospital, one of which did not report mortality and reported a 'poor outcome' by GOS criteria in an unconventional way. Substantial missing data in a key trial meant that in meta-analysis we report 'best-case' and 'worst-case' estimates alongside available case analysis. In no scenario did we discern a clear difference between treatments for either mortality or poor neurological outcome. Due to variation in modes of drug administration (including whether it followed or did not follow cerebrospinal fluid (CSF) drainage, as well as different follow-up times and ways of reporting changes in intracranial pressure, as well as no uniform definition of 'uncontrolled intracranial pressure', we did not perform meta-analysis for this outcome and report results narratively, by individual trial. Trials tended to report both treatments to be effective in reducing elevated intracranial pressure but that hypertonic saline had increased benefits, usually adding that pretreatment factors need to be considered (e.g. serum sodium and both system and brain haemodynamics). No trial provided data for our other outcomes of interest. We consider evidence quality for all outcomes to be very low, as assessed by GRADE; we downgraded all conclusions due to imprecision (small sample size), indirectness (due to choice of measurement and/or selection of participants without traumatic brain injury), and in some cases, risk of bias and inconsistency. Only one of the included trials reported data on adverse effects; a rebound phenomenon, which was present only in the comparator group (mannitol). None of the trials reported data on pulmonary oedema or acute renal failure during treatment. On the whole, trial authors do not seem to have rigorously sought to collect data on adverse events.
AUTHORS' CONCLUSIONS
This review set out to find trials comparing hypertonic saline to a potential range of other intracranial pressure-lowering agents, but only identified trials comparing it with mannitol or mannitol in combination with glycerol. Based on limited data, there is weak evidence to suggest that hypertonic saline is no better than mannitol in efficacy and safety in the long-term management of acute traumatic brain injury. Future research should be comprised of large, multi-site trials, prospectively registered, reported in accordance with current best practice. Trials should investigate issues such as the type of traumatic brain injury suffered by participants and concentration of infusion and length of time over which the infusion is given.
Topics: Brain Injuries; Brain Injuries, Traumatic; Glasgow Outcome Scale; Humans; Intracranial Hypertension; Intracranial Pressure; Randomized Controlled Trials as Topic; Saline Solution, Hypertonic
PubMed: 31978260
DOI: 10.1002/14651858.CD010904.pub3 -
The Cochrane Database of Systematic... Aug 2023Bullous pemphigoid (BP) is the most common autoimmune blistering disease. Oral steroids are the standard treatment. We have updated this review, which was first... (Review)
Review
BACKGROUND
Bullous pemphigoid (BP) is the most common autoimmune blistering disease. Oral steroids are the standard treatment. We have updated this review, which was first published in 2002, because several new treatments have since been tried.
OBJECTIVES
To assess the effects of treatments for bullous pemphigoid.
SEARCH METHODS
We updated searches of the following databases to November 2021: Cochrane Skin Specialised Register, CENTRAL, MEDLINE, and Embase. We searched five trial databases to January 2022, and checked the reference lists of included studies for further references to relevant randomised controlled trials (RCTs).
SELECTION CRITERIA
RCTs of treatments for immunofluorescence-confirmed bullous pemphigoid.
DATA COLLECTION AND ANALYSIS
At least two review authors, working independently, evaluated the studies against the review's inclusion criteria and extracted data from included studies. Using GRADE methodology, we assessed the certainty of the evidence for each outcome in each comparison. Our primary outcomes were healing of skin lesions and mortality.
MAIN RESULTS
We identified 14 RCTs (1442 participants). The main treatment modalities assessed were oral steroids, topical steroids, and the oral anti-inflammatory antibiotic doxycycline. Most studies reported mortality but adverse events and quality of life were not well reported. We decided to look at the primary outcomes 'disease control' and 'mortality'. Almost all studies investigated different comparisons; two studies were placebo-controlled. The results are therefore based on a single study for each comparison except azathioprine. Most studies involved only small numbers of participants. We assessed the risk of bias for all key outcomes as having 'some concerns' or high risk, due to missing data, inappropriate analysis, or insufficient information. Clobetasol propionate cream versus oral prednisone Compared to oral prednisone, clobetasol propionate cream applied over the whole body probably increases skin healing at day 21 (risk ratio (RR 1.08, 95% confidence interval (CI) 1.03 to 1.13; 1 study, 341 participants; moderate-certainty evidence). Skin healing at 21 days was seen in 99.8% of participants assigned to clobetasol and 92.4% of participants assigned to prednisone. Clobetasol propionate cream applied over the whole body compared to oral prednisone may reduce mortality at one year (RR 0.73, 95% CI 0.53 to 1.01; 1 study, 341 participants; low-certainty evidence). Death occurred in 26.5% (45/170) of participants assigned to clobetasol and 36.3% (62/171) of participants assigned to oral prednisone. This study did not measure quality of life. Clobetasol propionate cream may reduce risk of severe complications by day 21 compared with oral prednisone (RR 0.65, 95% CI 0.50 to 0.86; 1 study, 341 participants; low-certainty evidence). Mild clobetasol propionate cream regimen (10 to 30 g/day) versus standard clobetasol propionate cream regimen (40 g/day) A mild regimen of topical clobetasol propionate applied over the whole body compared to the standard regimen probably does not change skin healing at day 21 (RR 1.00, 95% CI 0.97 to 1.03; 1 study, 312 participants; moderate-certainty evidence). Both groups showed complete healing of lesions at day 21 in 98% participants. A mild regimen of topical clobetasol propionate applied over the whole body compared to the standard regimen may not change mortality at one year (RR 1.00, 95% CI 0.75 to 1.32; 1 study, 312 participants; low-certainty evidence), which occurred in 118/312 (37.9%) participants. This study did not measure quality of life. A mild regimen of topical clobetasol propionate applied over the whole body compared to the standard regimen may not change adverse events at one year (RR 0.94, 95% CI 0.78 to 1.14; 1 study, 309 participants; low-certainty evidence). Doxycycline versus prednisolone Compared to prednisolone (0.5 mg/kg/day), doxycycline (200 mg/day) induces less skin healing at six weeks (RR 0.81, 95% CI 0.72 to 0.92; 1 study, 213 participants; high-certainty evidence). Complete skin healing was reported in 73.8% of participants assigned to doxycycline and 91.1% assigned to prednisolone. Doxycycline compared to prednisolone probably decreases mortality at one year (RR 0.25, 95% CI 0.07 to 0.89; number needed to treat for an additional beneficial outcome (NNTB) = 14; 1 study, 234 participants; moderate-certainty evidence). Mortality occurred in 2.4% (3/132) of participants with doxycycline and 9.7% (11/121) with prednisolone. Compared to prednisolone, doxycycline improved quality of life at one year (mean difference 1.8 points lower, which is more favourable on the Dermatology Life Quality Index, 95% CI 1.02 to 2.58 lower; 1 study, 234 participants; high-certainty evidence). Doxycycline compared to prednisolone probably reduces severe or life-threatening treatment-related adverse events at one year (RR 0.59, 95% CI 0.35 to 0.99; 1 study, 234 participants; moderate-certainty evidence). Prednisone plus azathioprine versus prednisone It is unclear whether azathioprine plus prednisone compared to prednisone alone affects skin healing or mortality because there was only very low-certainty evidence from two trials (98 participants). These studies did not measure quality of life. Adverse events were reported in a total of 20/48 (42%) participants assigned to azathioprine plus prednisone and 15/44 (34%) participants assigned to prednisone. Nicotinamide plus tetracycline versus prednisone It is unclear whether nicotinamide plus tetracycline compared to prednisone affects skin healing or mortality because there was only very low-certainty evidence from one trial (18 participants). This study did not measure quality of life. Fewer adverse events were reported in the nicotinamide group. Methylprednisolone plus azathioprine versus methylprednisolone plus dapsone It is unclear whether azathioprine plus methylprednisolone compared to dapsone plus methylprednisolone affects skin healing or mortality because there was only very low-certainty evidence from one trial (54 participants). This study did not measure quality of life. A total of 18 adverse events were reported in the azathioprine group and 13 in the dapsone group.
AUTHORS' CONCLUSIONS
Clobetasol propionate cream applied over the whole body is probably similarly effective as, and may cause less mortality than, oral prednisone for treating bullous pemphigoid. Lower-dose clobetasol propionate cream applied over the whole body is probably similarly effective as standard-dose clobetasol propionate cream and has similar mortality. Doxycycline is less effective but causes less mortality than prednisolone for treating bullous pemphigoid. Other treatments need further investigation.
Topics: Humans; Azathioprine; Prednisone; Clobetasol; Pemphigoid, Bullous; Doxycycline; Methylprednisolone; Dapsone; Niacinamide
PubMed: 37572360
DOI: 10.1002/14651858.CD002292.pub4