Background: Artificial intelligence (AI) is rapidly reshaping clinical decision-making, yet it introduces ethically salient risks including bias, automation bias, opacity, privacy threats, data drift, adversarial vulnerability, and unclear accountability. Objective: To define and operationalize Clinical Cyberbioethics as an integrative framework for AI-mediated clinical decision-making, synthesizing ethical principles, governance requirements, and patient-facing digital rights proposed across the literature. Methods: We conducted a mapping and narrative review with SWiM-guided thematic synthesis covering January 2015 to March 2025, and documented identification, screening, and inclusion decisions using a PRISMA 2020–adapted flow diagram. Records were identified from an author-curated reference set compiled during the stated search period; screening and selection are reported in Fig. 1 and Supplementary Files 1–4. Results: Records identified (n = 54); duplicates removed (n = 11); records screened (n = 43); records excluded at title/abstract (n = 1); full-text sources assessed for eligibility (n = 42); studies included in synthesis (n = 42); full-text exclusions with reasons (n = 0). The synthesis yielded operational domains spanning decision authority, explainability/contestability, bias and equity safeguards, lifecycle governance and accountability, and proportional data governance and privacy safeguards. Conclusions: Clinical Cyberbioethics provides a practical framework to guide trustworthy AI-mediated clinical decision-making by integrating ethical principles with governance and patient rights.

Universal health coverage (UHC) requires simultaneous gains in service coverage and financial protection, yet these dimensions are often analysed separately. We conducted a secondary ecological panel study using two public indicator files (UHC-SCI and UHC-FH40). Records were first harmonized through a document-oriented non-relational workflow that preserved irregular subgroup structures before flattening to a country-year panel. The analytic sample comprised 981 observations from 159 countries/economies between 2000 and 2023. We estimated pooled trends, correlations, country and two-way fixed-effects models with clustered standard errors, a random-intercept model, a generalized estimating equation, domain-specific models, wealth- and urbanization-based inequality metrics, beta-convergence models, and country typologies. Mean service coverage rose from 58.9 to 74.1, whereas mean financial hardship fell from 24.0% to 17.3%. Each 1-point increase in the service coverage index was associated with a 0.441 percentage-point reduction in hardship (95% CI: -0.707 to -0.175; p = 0.001). The mean poorest-richest and rural-urban hardship gaps were 53.7 and 12.5 percentage points. Low-income settings showed the steepest inequities and the strongest negative slope. UHC progress therefore remains incomplete unless service expansion, financial protection, and inequality monitoring are interpreted together.

Postoperative pneumonia synthesis requires compatible outcomes, adjustment sets, and independent populations. This study examines how eligibility and modelling decisions affect the interpretation of perioperative vulnerability. A targeted, non-exhaustive review used 159 references and an inventory of 58 candidate reports. Four hip-fracture cohorts contributed to exploratory meta-analyses using restricted maximum likelihood and modified Hartung–Knapp inference. Nine analyses were checked using two Python implementations; eight leave-one-out analyses and twelve crude-to-adjusted comparisons were examined. Eighteen AI-assisted QUIPS domain judgments covered three cohorts; independent human reconciliation was not completed. The quantitative dataset comprised 6223 patients and 461 pneumonia events. Under permissive eligibility, the adjusted odds ratio for age was 1.47 per decade (95% confidence interval 1.06–2.02). Strict eligibility retained two cohorts, below the protocol-specific pooling threshold. Removing Tian from the unadjusted male-sex analysis changed the pooled odds ratio from 1.33 to 1.02 (0.50–2.07). The implementations agreed within 0.000001. The audit identified discordant matched-sample denominators in Shin. Other complications remained separate outcomes. Numerical reproducibility does not establish absence of bias. The inventory does not establish 57 independent eligible cohorts, and these findings cannot be reported as a completed systematic review.

Objective: To characterize the structural fragility of installed health-service capacity in Barranquilla, Atlántico, Colombia, using absolute capacity, supply concentration, reserve or transitory capacity, and service-line clinical sensitivity as structural-risk dimensions. Methods: An ecological health-services study was conducted using a local installed-capacity dataset traceable to the Colombian Special Registry of Health Service Providers and SISPRO, together with two contextual World Bank series for Colombia: physicians per 1,000 population and premature mortality from noncommunicable diseases. Traceable data cleaning, functional normalization, separation of baseline versus transitory capacity when allowed by the source fields, and exploratory estimation of a relative structural fragility proxy index using a normalized Poisson-type transformation were performed. This index was interpreted exclusively as a comparative structural-fragility ranking and not as an observed probability of saturation. Results: The analytical capacity of the Barranquilla node included 5,397 installed capacity slots. Adult ICU accounted for 707 slots and neonatal ICU for 160. Reserve capacity was low in neonatal ICU (2.5%) and higher in adult ICU (32.2%). The largest service lines were adult general hospitalization, adult ICU, and pediatric general hospitalization, whereas the highest relative structural fragility was observed in low-scale and highly concentrated services, including burn care, acute mental health, and selected highly specialized lines. Conclusion: Barranquilla has a broad but markedly heterogeneous structural health-service capacity network. The critical pattern is not determined only by the absolute number of slots, but by the interaction between limited capacity, high concentration, low stable reserve, and clinical sensitivity. The evidence generated is structural and should not be interpreted as observed occupancy, real-time saturation, or operational collapse.

Background: Surgical hand preparation, patient skin antisepsis, draping readiness and sterile-table set-up are often taught as separate rituals, although delays and failures arise at their interfaces. Evidence also spans microbiology, infection prevention, human factors and education, making it difficult to define what learners should time, observe, communicate and re-check. Methods: A rapid scoping review mapped global indexed evidence on three sterile-start processes: surgical hand preparation; the transition from skin antisepsis to draping or incision; and sterile-field or instrument-table set-up. Web of Science and Scopus were searched from inception to 2 August 2026. Records were deduplicated and subjected to a high-recall rule-assisted primary screen. All candidate records underwent eligibility verification, and the full text of every included source was confirmed, verified and analysed by the author team. Data were charted by domain, design, specialty, educational outcome and first indexed affiliation region. Findings were integrated into the proposed Surgical Timing and Aseptic Readiness Tool for the Operating Room (START-OR). Results: Of 5747 imported records, 298 duplicates were removed, 747 candidates underwent structured eligibility verification and 143 full-text evidence sources were included after author-team confirmation and analysis. The map comprised 30 education, assessment or implementation sources; 44 surgical hand-preparation sources; 61 skin-preparation sources; and eight sterile-field or table sources. Evidence did not support one universal minute value across products and contexts. Instead, hand preparation required product-specific technique, complete coverage, specified duration and drying; skin preparation required complete visible drying, absence of pooling and fire-risk control before draping; and sterile-field management required minimizing unnecessary open exposure and re-validating integrity after delay. Educational studies identified recurrent gaps in timing accuracy, coverage, feedback, monitoring and shared responsibility. START-OR therefore separates team, patient and field clocks, then reunites them at a voiced sterile-start pause and seven readiness gates. Conclusions: Perioperative education should replace memorized universal times with observable readiness criteria, interprofessional verification, simulation of delays and feedback linked to patient-safety outcomes. START-OR is a proposed, unvalidated curriculum and quality-improvement model that requires prospective multicentre testing before clinical adoption.

Universal health coverage requires simultaneous progress in service coverage and financial protection, yet these dimensions are commonly assessed separately. We conducted a secondary ecological panel analysis using official World Health Organization/World Bank indicators. A document-oriented workflow preserved irregular subgroup structures before conversion to an unbalanced country–year panel comprising 981 observations from 159 countries/economies between 2000 and 2023. The primary inferential specification was a two-way fixed-effects model with country and year effects and country-clustered standard errors; a country fixed-effects model with a linear time trend, a random-intercept model, and generalized estimating equations were complementary robustness analyses. We also examined service domains, wealth- and rural-urban inequalities, beta convergence, and exploratory country typologies. Mean service coverage rose from 58.9 in 2000 to 74.0 in 2023, whereas mean financial hardship fell from 24.0% to 17.3%. In the primary model, each one-point increase in service coverage was associated with a 0.441-percentage-point reduction in hardship (95% confidence interval, −0.707 to −0.175; p = 0.001). Mean poorest–richest and rural–urban hardship gaps were 53.7 and 12.5 percentage points, respectively. The inverse association was largest in low-income settings, where socioeconomic gradients were also steepest. Service coverage gains should therefore not be interpreted as sufficient evidence of equitable universal health coverage; monitoring systems and financing reforms should jointly target access, financial protection, and distributional gaps.