Although microprisms have become an important medical means of strabismus treatment, related research concerning the design, fabrication, and testing of microprismatic glasses for preventing eyestrain has rarely been reported. In this study, the structure of microprismatic glasses for preventing eyestrain related to using electronic monitors, including computers and mobile phones, is introduced. A designing theory of anti-fatigue glasses with microprisms is developed. The fabrication technique and the process are described, and the performances of the fabricated microprisms are characterized. Finally, a compact testing system for the measurement of prismatic diopter is designed and constructed. This measuring system can be used not only for Fresnel microprisms, but also for other types of prisms. The measured results agree with our calculations. Although this study is focused on optimizing the objective prismatic diopter for anti-fatigue microprismatic glasses, 2.0–3.0 prismatic diopters (Δ) for each eye in the anti-fatigue glasses are suggested according to our experience on strabismus treatments. The clinical research for patients using the developed anti-fatigue glasses will be fully implemented in our further research to confirm the optimal subjective prismatic value. Anti-fatigue glasses can be very useful for different applications such as reading books, making precise technological operations, and working with the electronic monitors of computers and mobile phones. The last case is the most important, because many people now work with computers for two or more hours per day; this is the normal regime for working, yet is a very intensive strain on the eyes ,2,3.Glasses for using computers have about 60 percent of the magnifying power of general-use glasses, because computer screens are usually positioned at about 600 mm from the user’s eyes . The simDuring long periods of work with computers or mobile phones, the eye muscles constantly tense up and incline the sight lines from both eyes to a single point on the screen. It is well known that for all work connected with tension of the muscular apparatus of the eye, the probability exists of an excessive increase in intraocular pressure ,11,12. TBesides the effects on the eyeball itself, to focus the beam from a close object onto the retina the two oblique muscles encircling the eyes are under the compressive force . In addiFrom the above, it can be concluded that it is only possible to avoid eye fatigue during long-term work close to the eyes if the conditions can be created in which the examined near object is treated as far away. One possible method is to change the direction of the light beams from the near objects by introducing an optical element and allowing the light from the near objects to reach the eyes via parallel beams. The first efforts to reduce the stress of eye accommodation can be traced back to the late 1950s, when glasses based on the combination of a refractive lens and a prism were proposed by Utekhin ,19. The In our previous research, we investigated the use of Fresnel prisms for strabismus treatment in children ,22. A toTo the best of our knowledge, the design, fabrication, and testing of glasses with microprisms for preventing eyestrain have rarely been reported until now. In this study, the structure of microprismatic glasses for preventing eyestrain is introduced. The design theory of anti-fatigue glasses with microprisms is developed. The fabrication technique and process are described, and the performance of the fabricated microprisms is analyzed and discussed. Finally, a compact measuring system for the prismatic diopter of the prisms is designed and built. Clinical research for patients using anti-fatigue glasses is also started. It has been verified by several research participants that utilization of the proposed anti-fatigue microprismatic glasses can improve vision and relieve eyestrain.The function of preventing eyestrain is fulfilled by microprisms, which can change the direction of the light beams from a near object to almost parallel; therefore, the microprism is the main device in the anti-fatigue glasses. For cases without ametropia, the anti-fatigue glasses can simply be spectacles with two microprism components. However, this simple structure has the disadvantage of tending to be contaminated by moisture and dust, as stated previously. A modified structure combines a plain piece of glass and a microprism as a hermetic construction with the microrelief inside similar to the one shown in For some advanced users, the anti-fatigue glasses can be designed using a more complex structure; one piece of microprism can be divided into two zones as a two-zone microprism, similar with the idea in ,23. The Although the anti-fatigue microprismatic glasses are similar to the glasses used for the therapeutic treatment of strabismus, the optimal prismatic diopter, the fabrication, and the performance characterization are still open questions.To create anti-fatigue prismatic glasses the first issue is the optical center of the glasses, which is the center-to-center distance between the eyes. For cases without ametropia, the optical center is not important as the prism on the glasses is placed perpendicular to the relief strokes. However, in the presence of refractive components the features of a patient’s eyes should be considered, including the center-to-center distance. Generally, the optical center can be decided by the pupil distance data.The most important parameter for anti-fatigue prismatic glasses is the prismatic diopter. Utekhin’s glasses were made with a total prism strength of 7.5 diopters (Δ) for both eyes , i.e., 3The standard viewing angle of objects on a computer screen at a viewing distance of 350 mm is about 5.06–5.22° for a pupil distance of 62–64 mm, which is a typical value of pupil distance for adults. The calculated values of the necessary prismatic diopter, according to the geometrical optics and the characterization of optical materials with the refractive index of 1.46–1.52, are shown in However, with the anti-fatigue microprismatic glasses the light beams are focused on the screen not only according to the prism action, but also due to the fusion provided by the human brain. Therefore, the optimal prismatic diopter should be defined as two different assessed values: the objective diopter and the subjective diopter. The objective prismatic diopter can be calculated according to geometrical optics; this is the subject of the current study. For the subjective prismatic diopter it is necessary to carry out detailed clinical studies with patients, which is beyond the scope of this paper, although we have already started the related clinical study. We think that the optimal value of the prismatic diopter in the anti-fatigue glasses can be obtained only after these precise clinical investigations. In our opinion, this prismatic diopter value should be in the range of 2.0–3.0 Δ for each eye.W0 stands for the length of the bottom of the prism unit , L0 stands for the height of the prism unit, α and β are the angles defining the prism shape, and n1 and n2 are the refractive indexes of the microprism and the surroundings, respectively. The parameter that decides the prismatic diopter of the prism is α, and β is related to the technological requirement of reducing stray light. According to our previous experience [β should be set as 2°–3° for reducing stray light.The main problem with microprisms in anti-fatigue glasses is the diffraction at the microrelief that affects the visual images. The other physical disadvantage is the chromatism of white light . For micperience , β shoulφ stands for the diffractive angle, λ is the wavelength of incident radiation, S0 is the width of the diffractive slit when a single prism unit is considered, I0 is the initial optical intensity, and For the consideration of diffraction phenomena, the microprism can be treated as a blazed grating and the visual image through it will be affected by the single-slit diffraction from every single prism unit and the multiple-diffractive-beam interference from the whole microprism. The intensity of diffraction from a single prism unit can be described as :(1)Id=I0N stands for the number of prism units, D1 is the width of the interference slit between two diffractive beams from prism units, and The intensity of multiple-diffractive-beam interference from the whole microprism can be described as :(2)Ii=I0Therefore, the optical intensity through the microprism can be described as:β = 0, we get When When W0 ~ 100–200 μm the expansion of optical images after the microprism became very large (Δγ ~ 0.2°–0.4°) compared with the prism deflection angle γ ≈ 5.2° for a pupil distance of 64 mm and an observation distance of 350 mm. The smallest resolution angle of a normal eye is usually 0.017°, therefore W0 should be around 600 μm. The value N is defined by the size of the microrelief area, which is illuminated by a laser beam. In our calculations, if the typical value of the pitch width is W0 = 600 µm, then N = 4 is a reasonable value that corresponds to the diameter of the laser beam. The working wavelength λ is chosen as 6328 nm.According to the equations derived above, the simulations are performed and the obtained results are shown in γ, W0 = 530 μm, the corresponding angle is just 0.017°; therefore, the pitch width W0 should be larger than 530 μm. However, if W0 is too large the anti-fatigue microprismatic glasses will be inconvenient for wearing due to visible streaking of the images.W0 = 600–800 μm. For lower pitches the microprisms would distort the observed optical images. For larger pitches the anti-fatigue glasses would not be attractive from a cosmetic viewpoint because the structures of the microrelief become clearly visible.Thus, the optimal microrelief pitch width for the anti-fatigue microprismatic glasses is better set as Microprism is a kind of micro-structured optical component. Usually, it can be fabricated by micro-fabricated technologies such as lithography, etching, and laser direct writing. However, for the case of microprisms the micro-fabricated techniques mentioned above are infeasible because there are slant angles related with the structure of microprisms. Therefore, a stamper with specially designed prismatic microrelief was manufactured by ultrahigh precision machining for meeting the requirements, including the accuracy of structure parameters and the optical surface of a microprism. Moreover, considering the feasibility of the mass production of microprisms, optical thermoplastic materials are the most suitable materials for actually manufacturing the microprisms by hot pressing or injection molding technology.According to their material characteristics, polymethylmethacrylate (PMMA), polyvinylchloride (PVC), and polycarbonate (PC) are suitable materials for fabricating the anti-fatigue microprismatic elements for glasses; of these PMMA is the best, because it has a smaller chromatic dispersion, a suitable refractive index (~1.492), higher light transmissivity (90–92%), and better thermoplasticity . The disα of the prism unit is increasingly smaller. The microprisms were fabricated by hot pressing technology in this study. A pressing stamper was designed according to the results of the theoretical simulation of the microprism parameters by the method introduced above. The prismatic microrelief was then fabricated upon an extra super duralumin alloy (V-95) at the Institute for Information Recording, the National Academy of Sciences of Ukraine . The fabThe PMMA microprisms were fabricated by a servo hot pressing machine , which has a cooling system for a better molding performance. A laser-cutting machine used for obtaining a specific shape of PMMA substrate was also used. First, the pressing stamper was put on the cut and then the cleaned PMMA substrate; together they were placed in the lower mold of the servo hot pressing machine and preheated to 60–70 °C. Then, the upper mold of the servo hot pressing machine was heated to 150–165 °C and pressed on the pressing stamper with a force of 100–200 kg for 30 s. Afterwards, the cooling system was turned on in order to decrease the temperature to 60 °C under the condition of maintaining constant pressure. Finally, the cooling system was turned off, the pressure was released, and we obtained the PMMA microprismatic element. W0) and the height of the prism unit (L0). On the basis of the measured results we can also characterize the surface finish (aR) of the fabricated microprism by numerical analysis. The surface finish (aR) can be calculated by the contour arithmetic mean deviation:m stands for the sampling number within a sampling length, i is a positive integer, and h is the absolute value between the specific point on the contour and the reference point on the datum plane. To inspect the quality and the accuracy of the fabricated microrelief on the microprism, a scanning electron microscope  and a stylus profiler system  were used to measure the structure parameters of the microrelief. lE) of 10 mm on the screen with a distance (LE) of 1000 mm from the prism to the screen corresponds to a prismatic diopter of 1.0 prismatic diopter (Δ); therefore, for any deflection angle γ the prismatic diopter can be easily calculated as Δ = 100 (lE /LE) [The prismatic diopter is the most important parameter of the developed anti-fatigue microprismatic glasses. However, to the best of our knowledge there is no measuring system for testing the prismatic diopter of a prism or a microprism. Although we can indirectly obtain the prismatic diopter of a prism by measuring its structural parameters, it is necessary to develop a measuring system for directly testing the prismatic diopter of prisms. According to the theory introduced in (lE /LE) .γmax) corresponding to the largest prismatic diopter as the reference value; therefore, the rotation angle should be increased depending on the increase in prism strength for prisms with smaller Δ.To minimize the size of the measuring setup, it is possible to use a special scheme with a rotating frame with microprisms. We take the deflection angle . For all other microprisms, the values of the rotation angle (θ) should be obtained for which the deflection angle γ = γ30 = 16.69924°, depending on the prismatic diopter for all tested microprisms. The dependence of the prismatic diopter on the rotation angle θ was investigated [A microprism with 30.0 Δ has a deflection angle of stigated ,22 and sLE to 250 mm and in turn reduce the size of the measuring system. The adjustable diaphragm in front of the first reflector must be set in the direction of γ30, because we took the largest prismatic diopter as the reference value. In addition, the laser must exit horizontally and incident on the flat surface without microrelief to avoid measurement error; moreover, the element holder can only be rotated in one direction to improve the measuring accuracy. A compact measuring system for obtaining the prismatic diopter of prisms was developed using the left part of the curves in We measured the five fabricated samples using the developed compact measuring system; the measured results are shown in γ on the rotation angle θ for all prisms with prismatic diopters of 1.0–30.0 Δ. From the calculation results, it follows that for very low strengths of 0.5–2.0 Δ it is not possible to achieve the deflection angle of γ30 = 16.69924° by rotation of the prism, because the reflection angle of the laser beam inside the prisms exceeds the critical reflection angle in this case [γ30 = 16.69924°. For example, for a prism with a prismatic diopter of 1.0 Δ the largest possible rotation angle is θ = 77.76°, which corresponds to a deflection angle (γ) of 10.66°. For a prism with a prismatic diopter of 2.0 Δ the largest possible rotation angle θ = 72.598°, which corresponds to a deflection angle (γ) of 14.94°.We calculated and analyzed the dependence of the value of his case . It meanThus, the measuring system developed in this study can only efficiently measure the prismatic diopter of prisms within a range of 3.0–30.0 Δ. We measured microprisms with prismatic diopters from 3.0 Δ to 30.0 Δ using the developed measuring system and compared the measured results with our theoretical calculations, finding that they agreed well (see The possibility of using Fresnel microprismatic glasses for preventing eyestrain was thoroughly investigated. Firstly, the construction of the anti-fatigue glasses was introduced. Next, the design theory of the anti-fatigue glasses with microprisms was developed and the simulations were performed, which demonstrated the effectiveness of the proposed anti-fatigue microprismatic glasses.Subsequently, the fabrication technique and the process were described and the general quality of the fabricated microprisms was tested by an optical microscope. In addition, the structural parameters of the microprisms were measured by a stylus profiler system and a scanning electron microscope. The measured results demonstrated the fine accuracy and the optical surface of the fabricated microprisms.Finally, a compact testing system for measuring the prismatic diopter of the prisms was designed and built. The prismatic diopters of the fabricated microprisms were measured by the developed measuring system and the measured results agreed with the calculations. The compact measuring system we developed for the measurement of the prismatic diopter can be used for any kind of prism.Although clinical research for patients using anti-fatigue glasses has already started, in this paper we focused on the optical theory and the techniques related to the fabrication and the measurement of the microprisms. The optimal value of the prismatic diopter of prisms in the anti-fatigue microprismatic glasses can only be obtained after precise clinical investigations. In our opinion, this value should be within the range of 2.0–3.0 prismatic diopters for each eye to help the brain achieve the total necessary convergence of the eyes for viewing near objects.


Ovejero et al. show that cooling resting or relaxed rat heart muscle releases myosin motors from the helical folded conformation that inhibits actin interaction in physiological conditions. Myosin filament–based regulation supplements actin filament–based regulation to control the strength and speed of contraction in heart muscle. In diastole, myosin motors form a folded helical array that inhibits actin interaction; during contraction, they are released from that array. A similar structural transition has been observed in mammalian skeletal muscle, in which cooling below physiological temperature has been shown to reproduce some of the structural features of the activation of myosin filaments during active contraction. Here, we used small-angle x-ray diffraction to characterize the structural changes in the myosin filaments associated with cooling of resting and relaxed trabeculae from the right ventricle of rat hearts from 39°C to 7°C. In intact quiescent trabeculae, cooling disrupted the folded helical conformation of the myosin motors and induced extension of the filament backbone, as observed in the transition from diastole to peak systolic force at 27°C. Demembranation of trabeculae in relaxing conditions induced expansion of the filament lattice, but the structure of the myosin filaments was mostly preserved at 39°C. Cooling of relaxed demembranated trabeculae induced changes in motor conformation and filament structure similar to those observed in intact quiescent trabeculae. Osmotic compression of the filament lattice to restore its spacing to that of intact trabeculae at 39°C stabilized the helical folded state against disruption by cooling. The myosin filament structure and motor conformation of intact trabeculae at 39°C were largely preserved in demembranated trabeculae at 27°C or above in the presence of Dextran, allowing the physiological mechanisms of myosin filament–based regulation to be studied in those conditions. The contraction of striated muscle is generated by relative sliding of actin-containing thin filaments and myosin-containing thick filaments driven by myosin motors and is initiated by an intracellular calcium transient elicited by an action potential . CalciumThick filament–based regulation is thought to be disrupted by mutations in thick filament proteins associated with heart disease , so it rIn resting or relaxed mammalian skeletal muscle, myosin motors can be released from the helical folded conformation by cooling below physiological temperature , and thiDisruption of the helical order of the thick filament has been also reported in relaxed demembranated cardiac trabeculae cooled from 25°C to 5°C . More reHere, we used x-ray diffraction to characterize the structural changes in the thick filament associated with cooling intact quiescent trabeculae from 39°C to 7°C. We show that the myosin motors leave the helical folded state on cooling, and the periodicity of the thick filament backbone increases. Because cooling intact trabeculae alters their intracellular ionic composition, we also characterized the effect of cooling on thick filament structure in relaxed demembranated cardiac trabeculae, in which the ionic composition of the solution bathing the myofilaments is controlled. Addition of the osmotic agent Dextran T500 (3% wt/vol) to the relaxing solution restored the filament lattice spacing to that in intact muscle cells at 39°C and stabilized the folded helical conformation of the myosin motors against the effects of cooling.Rattus norvegicus, male, 6–8 wk old) were supplied by Charles River Laboratories and hosted at the animal house of the Bio-Medical Facility of the European Synchrotron Radiation Facility (ESRF). On the day of the experiment, the rats were euthanized by cervical dislocation after sedation with isoflurane in compliance with the UK Home Office Schedule 1 and European Union regulation (directive 2010/63), followed by a confirmation method. The heart was rapidly excised and cannulated via the ascending aorta and retrogradely perfused with a modified Krebs-Henseleit buffer  equilibrated with carbogen . The pH of the Krebs solution decreased from ∼7.5 at 36°C to ∼7.3 at 7°C. Single unbranched trabeculae were dissected from the right ventricle under a stereomicroscope and were selected to be used as intact or demembranated samples.The cardiac trabeculae were prepared as previously described . Briefly2 without BDM at 27.6 ± 0.1°C . Two mica windows carrying platinum-stimulating electrodes were positioned as close as possible to the trabecula to minimize the x-ray path in solution, and the trabeculae were electrically paced at 1 Hz for at least 30 min before the start of the experiment to check the viability of the preparation. Continuous perfusion keeps the level of oxygen constant at the sample position. The temperature inside the trough was controlled by adjusting the temperature of the perfusate via a counter-current system and continuously monitored with a temperature probe. The speed of temperature change was very slow when warming/cooling due to the overall inertia of the system (10–20 min); therefore, cooling did not induce spontaneous contractures (2 (mean ± SD), and length was 2.64 ± 0.57 mm.For experiments on intact trabeculae, “scorpion-like” clips made of stainless-steel wire were mounted on the valve and ventricular wall ends of each trabecula. Trabecular length and cross-sectional area were measured after stretching to just above slack length. They were then mounted in the experimental plexiglass trough filled with the same buffer between the levers of a strain gauge force transducer and a motor  at slack length. The trabecula was then perfused with a buffer containing 1.4 mM CaClractures . Howeverractures ; thereforactures . In some2, 5 mM EGTA, 7 mM Na2ATP, 1 mM NaN3, and 50% glycerol) for up to 24 h. Aluminum T-clips were attached to the ends of the trabeculae, and they were washed in relaxing solution before mounting in a temperature-controlled multidrop apparatus (n = 4 trabeculae) between the levers of a strain gauge force transducer and the Aurora motor. Relaxing solution contained 25 mM Imidazole, 45 mM KPr, 6.89 mM MgAc2, 10 mM EGTA, 5.56 mM Na2ATP, 20 mM Na2-creatine phosphate, (pCa = −log [Ca2+] = 9), free [Mg2+] = 1.0 mM, ionic strength = 180 mM, and pH 7.1 at 23°C. In the same trabeculae, the osmotic agent Dextran T500 (3% wt/vol) was added to the relaxing solution to reduce the interfilament spacing to a value similar to that of intact trabeculae at body temperature. Just before the experiment Protease inhibitor cocktail P8340 (Sigma) and 2 mM dithiothreitol was added to all the solutions. The temperature of the multidrop trough was controlled by a Peltier system and recorded with a temperature probe. Due to the very small volume of each drop and the high thermal conductance of the aluminum base, the speed of temperature change was fast (<1 min). The reversibility of the effect of cooling on the meridional x-ray intensity distribution in demembranated trabeculae is shown in P2> has a temperature dependence similar to that of the x-ray signatures of the helical folded state of the myosin motors reported below, and its value at each temperature was not significantly affected when the pH of the relaxing solution was set to 7.1 at each temperature (2 (mean ± SD), and length was 1.63 ± 0.51 mm.For experiments on demembranated trabeculae, isolated trabeculae were demembranated for 20 min on ice in relaxing solution in the presence of BDM (25 mM) and Triton X-100 1% (vol/vol), then stored at −20°C in storage solution  and 50 µm  at the detector , Fit2D , and IgorPro . 2-D patterns were centered and aligned using the equatorial 1,0 reflections, then mirrored horizontally and vertically. For each trabecula, data obtained at the same temperature were added to increase signal-to-noise for the weakest reflections.−1 on either side of the equatorial axis (perpendicular to the trabecular axis), and the intensities and spacings of the 1,0 and 1,1 reflections were determined by fitting two Gaussian peaks in the region 0.019–0.071 nm−1 with the constraint d11 = d10/Equatorial intensity distributions were obtained by integrating from 0.0036 nm−1 on either side of the meridional axis . Background intensity distributions were fitted using a convex hull algorithm and subtracted; the small residual background was removed using the intensity distribution from a nearby region of the pattern containing no reflections. The interference components of the meridional reflections were determined by fitting multiple Gaussian peaks to the meridional intensity distribution from the following axial regions: M1-T1, 0.034–0.017 nm−1, four Gaussian peaks, the two lower angles with the same width for M1, the two higher angles with another width for T1; M2 cluster, 0.054–0.036 nm−1, four Gaussian peaks, the two lower angles with the same width for M2L, the two higher angles with another width for M2H  = 14.479 nm measured in intact trabeculae in diastole and in demembranated trabeculae at low calcium concentration (pCa 9) at 27°C  was integrated radially in the region between 0.063 and 0.023 nmI10 at 27°C.1-D profiles shown in Force, stimulus, fiber length change, and x-ray acquisition timing were collected and analyzed using custom-made software written in LabVIEW .d ∼14.5 nm. Each of the arrays contains 49 layers of three dimeric myosin molecules, or six motor domains. The thick filament was represented as a 1-D lattice of point diffractors with mirror symmetry about the filament midpoint, with the first lattice point in each half-filament at the half-bare zone (hbz), followed by the other 48 layers. hbz is an estimate of the average center of mass of the myosin motors; therefore, it can be used to determine changes in motor conformation; however it holds an uncertainty of ± 7.2 * i nm, where i is an integer (I(R) = [sin(NπRd)/sin(πRd)]2cos2(πRD), where R is the reciprocal space coordinate, d is the axial spacing of the myosin layers, N is the number of contributing myosin layers in a contiguous array in each half-filament, from a proximal layer np to a distal layer nd, while motors outside that array are isotropic and make no contribution to the reflection; D is the interference distance (ID) calculated as D = 2 * hbz + (np + nd − 2) * d (I(R) was convoluted with a Gaussian function with sigma ∼50 µm, representing the combined point-spread function of the x-ray beam and detector, and fitted to the experimental data  = [− 2) * d . The caldata see . The besfactor y , by minind in each half-filament and the axial periodicity d.The best-fit to the M2L, M2H, and M6 meridional intensity distributions at 39°C in intact trabeculae was calculated using a similar approach, but adjusting the maximum number of point-diffractors n = 4 trabeculae; X-ray diffraction patterns from intact quiescent trabeculae isolated from the right ventricle of rat hearts were collected at 39°C and SL 2.07 ± 0.05 µm  at 39°C and decreased slightly on cooling to 7°C, to 34.1 ± 0.3 nm (I11 and I10 reflections (I11/I10), often used as an indicator of the movement of mass within the myofilament lattice, increased on cooling from 0.26 ± 0.01 at 39°C to 0.42 ± 0.05 at 7°C  increased slightly (d10 increased to 41.7 ± 0.4 nm on cooling demembranated trabeculae to 9°C. I10 decreased on cooling (I11 (empty triangles) and I11/I10  from the folded and helically ordered myosin motors on the surface of the thick filament decreased on cooling in both intact quiescent trabeculae , the temperature at which the intensity changes were half-maximal, of ∼19°C in intact trabeculae and demembranated trabeculae in the presence of Dextran and of ∼24°C in the absence of Dextran  peak and small low-angle (la) and high-angle (ha) satellites (ID (IM3) in intact trabeculae decreased markedly on cooling . Its total intensity (IM6) decreased on cooling to 11°C, (SM6) at 39°C was 7.24 nm, roughly half of SM3 in those conditions  decreased on cooling with a transition temperature of ∼20°C  from the M-line and average periodicity 14.49 nm (d). The MyBP-C containing C-zone spans from layer 7 to 31 (green bars and light gray shade on the thick filament), the proximal or P-zone layers 1 to 7, and the distal or D-zone layers 31 to 49  so that its value at 22°C became similar to that in intact trabeculae (gray circle). The center-of-mass of the ordered motors moved ∼10 nm away from the M-line on cooling to 9°C  leads to an increase in the filament lattice spacing . At 38°Cilaments , the avents  is also similar in relaxing conditions at 7°C as at peak systolic force  had a lower transition temperature, ∼13°C. The axial periodicity of the myosin motors (SM3) had a distinct temperature dependence, first decreasing on cooling to a minimum at 10°C before recovering on further cooling. We return to the interpretation of this difference in a later section.The effects of cooling on the intensities of the ML1 layer line and the main meridional reflections were half-complete at a transition temperature of 18–20°C , althougThe interpretation of the effects of cooling intact quiescent trabeculae is complicated, however, by its effects on the intracellular ion concentrations. Cooling below 20°C inhibits the sodium-potassium pump with a consequent elevation of the intracellular sodium concentration and of calcium influx via the sodium-calcium exchanger . CardiacIML1; SM6  increases much more dramatically below 15°C  in these conditions, if the intrinsic axial periodicity of the filament is larger in the D-zone than the C-zone, as argued previously on the basis of the distinct changes in SM3 and SM6 following electrical stimulation of intact trabeculae at 27°C . Moreover, all these parameters are the same, within the resolution of the measurements, in quiescent trabeculae and in diastole in electrically paced trabeculae  of the myosin-based layer lines and meridional reflections at 38–39°C in intact and demembranated trabeculae in the presence of 3% Dextran T500.Click here for additional data file.Table S2lists x-ray parameters at 39°C and 7°C in intact quiescent and demembranated trabeculae in the absence and presence of Dextran.Click here for additional data file.Table S3reports the transition temperatures of x-ray parameters.Click here for additional data file.Table S4provides the model parameters for the best fit of the axial profile of the myosin (M) reflections.Click here for additional data file.


Oriented external electric fields are now emerging as “smart effectors” of chemical changes. The key challenges in experimentally studying electrostatic catalysis are (i) controlling the orientation of fields along the reaction axis and (ii) finely adjusting the magnitudes of electrostatic stimuli. Surface models provide a versatile platform for addressing the direction of electric fields with respect to reactants and balancing the trade-off between the solubility of charged species and the intensity of electric fields. In this mini-review, we present the recent advances that have been investigated of the electrostatic effect on the chemical reaction on the monolayer-functionalized silicon surfaces. We mainly focus on elucidating the mediator/catalysis role of static electric fields induced from either solid/liquid electric double layers at electrode/electrolyte interfaces or space charges in the semiconductors, indicating the electrostatic aspects is of great significance in the semiconductor electrochemistry, redox electroactivity, and chemical bonding. Herein, the functionalization of silicon surfaces allows scientists to explore electrostatic catalysis from nanoscale to mesoscale; most importantly, it provides glimpses of the wide-ranging potentials of oriented electric fields for switching on/off the macroscale synthetic organic electrochemistry and living radical polymerization. At present, increasing attention has been paid to use oriented electric fields as “smart reagents” to catalyze non-electrochemical reactions, namely, electrostatic catalysis . The conIt was not until 2016 that Coote et al. provided a proof of concept that an electric field could accelerate a carbon–carbon bond-forming reaction through scanning tunneling microscope-break junction (STM-BJ) experiments ; this navia homolytic cleavage of Si-H bonds at room temperature , describing that the extent of attractive and repulsive interactions between molecules  value] while dealing with redox reactions that occur in semiconductor electrodes . The curectively . These colecules  is direccreasing  or the screasing  is decrenterface , as wellnterface .via thermal hydrosilylation of 1,8-nonadiyne to produce alkyne-terminated monolayers at 160°C (4− (330 ± 2 mV) < NO3− (341 ± 9 mV) < ClO4− (406 ± 4 mV) < CF3SO3− (429 ± 1 mV) < PF6− (446 ± 8 mV). With the exception of HSO4−, the magnitude of the displacement in 4− compared with the other anions can be rationalized by the formation of a stabilizing H-bond with the 1,2,3-triazole ring. The displacement in i.e., oxoammonium–electrolyte anion interaction), and this force leads to a reduction in the thermodynamic cost for oxidizing TEMPO. Consequently, the electrochemical activity of TEMPO monolayers can be predictably manipulated by simply changing the electrolytes. Conversely, for immobilizing nitroxide radicals on the poorly doped Si(100) electrode, there is an internal electric field induced by the semiconductor space charge, which can be applied to drive the redox reaction of TEMPO contrathermodynamically. To our best knowledge, the kinetic barrier of depleted dark semiconductor electrodes can be effectively controlled, or even removed, upon illumination with different energies; in particular, the formed internal electric field in the space-charge region of semiconductors can predictably tune the electroactivity of redox species attached to the electrode surface, hence implying the electrostatic aspect in the semiconductor electrochemistry. Persistent organic free radicals have been extensively studied in chemistry and biology , and TEMat 160°C . The finat 160°C  between basicity . The stri.e., solvents, ion-pairing, and electrolytic concentration) in the electrolytes. Importantly, when the alkoxyamines are grafted on the silicon surface, one can control the number of fragments by changing the anodization time of applied positive biases, suggesting a versatile approach for generating in situ active nitroxides and carbocations used to regulate chemical synthesis. For example, we have covalently attached an alkoxyamine derivative to a Si(100) surface and applied an anodic bias to trigger its decomposition into a diffusive nitroxide radical and a surface-tethered carbocation (via a Cu(I)-catalyzed “click reaction”; as can be seen from in situ electrogenerated surface-confined carbocations immediately trap the preexisting nucleophiles , which is characterized by using the cyclic voltammogram with an obvious ferrocene redox signature. By changing the anodization time applied upon alkoxyamine monolayers, one is able to control the density of active carbocations and hence the coverage of a formed redox probe tagged monolayer. This proof-of-concept experiment expands the electrostatic effect to the regulation of organic electrosynthesis (Beyond utilizing electric fields to catalyze redox  or singlbocation . Differeynthesis .Since the first experimental evidence about the electrostatic catalysis of the C–C bond formation at individual-molecule interfaces , using oThe prospects of electrostatic catalysis is to establish the static electric field as a “smart reagent”—a catalyst or inhibitor  in a bro


Objectives: To investigate the association between birth weight and the risk of hypertension, and to examine the interaction between birth weight and the adult obesity index.Methods: We included 199,893 participants who had birth weight data and no history of hypertension at baseline (2006–2010) from the UK Biobank. A multivariate cubic regression spline was used to visually explore the dose-response relationship. Multivariate Cox proportional hazard regression models were used to calculate hazard ratios (HRs) and 95% confidence intervals (CIs).Results: We observed a nonlinear inverse association between birth weight and hypertension. The risk for hypertension decreased as birth weight increased up to approximately 3.80 kg. Compared with the participants with the fourth quintile of birth weight (3.43–3.80 kg), those with the first quartile of birth weight (<2.88 kg) were associated with a 25% higher risk of hypertension . In addition, the participants with birth weight <2.88 kg and body mass index ≥30 kg/m2 had the highest risk , as compared with those with birth weight between 3.43–3.80 kg and body mass index between 18.5–25.0 kg/m2. These associations were largely consistent in the stratified and sensitivity analyses.Conclusion: Our findings indicate that lower birth weight is nonlinearly correlated with higher risk of hypertension, and birth weight between 3.43–3.80 kg might represent an intervention threshold. Moreover, lower birth weight may interact with adult obesity to significantly increase hypertension risk. Fetal malnutrition, the primary indicator of which is low birth weight, can permanently alter organ structure and function in a way that predisposes the offspring to cardiovascular disease (CVD) in adulthood , 2. GrowAlthough some epidemiological studies have suggested a nonlinear inverse association between birth weight and hypertension –9, the iTherefore, for the current study we used data from the UK Biobank, a large prospective cohort study conducted in UK, to investigate the dose-response association of birth weight with hypertension and blood pressure. Moreover, we examined potential interactions between birth weight and the adult obesity index.The UK Biobank is a prospective cohort study recruiting half a million participants aged 37–73 years between 2006 and 2010 . The UK n = 147,989). We further excluded 151,078 individuals with missing birth weight data and 3,661 individuals with missing values on the main covariables. Finally, a total of 199,893 participants were included . Waist and hip circumferences were measured with a Seca200 measuring tape using standard procedures, and waist-to-hip ratio (WHR) was the ratio of waist circumference to hip circumference.Participants were asked to report their own birth weight . Standing height was measured using a Seca202 device. Body mass index (BMI) was defined as weight divided by height squared (mhttps://biobank.ndph.ox.ac.uk/showcase/showcase/docs/Bloodpressure.pdf" https://biobank.ndph.ox.ac.uk/showcase/showcase/docs/Bloodpressure.pdf). Means of SBP and DBP from two automated or two manual blood pressure measurements were calculated.Systolic blood pressure (SBP) and diastolic blood pressure (DBP) measurements were taken in a seated position after a few minutes of rest using an Omron 705 IT electronic blood pressure monitor. A manual sphygmometer was used if the standard automated device could not be employed. Further details of these measurements can be found in the UK Biobank online protocol  and Scottish morbidity records (Scotland). Date of death was obtained from death certificates held by the National Health Service Information Center  and the National Health Service Central Register Scotland (Scotland) . The priPerson-time of follow-up was calculated for each participant from the age in months at the return date of the baseline questionnaire (2006–2010) until the age in months at the date of first diagnosis of hypertension or the end of follow-up , whichever came first. Cox proportional hazards regression models with age as the time scale were used to analyze the association between birth weight and hypertension, reporting hazard ratio (HR) and 95% confidence intervals (CI).p-value < 0.05 denoting significant nonlinearity.A multivariate restricted cubic spline with four knots was used to visually explore the nonlinear association of birth weight with hypertension. We used a likelihood ratio test (LRT) to compare the model with only the linear term of birth weight to the model with both the linear and the cubic spline terms, with a n = 199,765). Birth weight was treated as a categorical (the quintiles) or continuous (per SD increment) variable.In the multivariate analysis of hypertension, we adjusted for age, sex, and early life events, including maternal smoking, breastfeeding, part of a multiple birth, and birthplace (Model 1). We also further adjusted for the Townsend deprivation index, college or university degree, BMI, summed metabolic equivalent of task hours per week (MET-hours/week) for physical activity, smoking status, alcohol intake frequency, intake of vegetables and fruit, family history of hypertension, and prevalence of diabetes (Model 2). Multivariate linear regression analysis was conducted to examine the associations between birth weight and SBP and DBP separately , maternal smoking , breastfeeding , physical activity , and smoking status . We tested the interaction between birth weight and each of the stratification variables using LRT, comparing a model with and without interaction terms.2 was treated as the reference group. P for interaction was assessed by a Wald test for the cross-product terms between birth weight and BMI (continuous). Sensitivity analyses were performed by excluding individuals who were part of multiple births .In the joint analysis of birth weight and BMI, the fourth quintile of birth weight (3.43–3.80 kg) and BMI in adulthood between 18.5–25.0 kg/mP < 0.05 was defined as statistically significant.All statistical tests were two-sided, and SAS version 9.4 (SAS Institute) was used for all analyses. During a median follow-up of 8.8 (interquartile range: 8.1–9.4) years, 12,333 hypertension events occurred among 199,893 participants. The basic characteristics of participants according to the quintile of birth weight are shown in p for nonlinearity = 0.0004).A nonlinear inverse association between birth weight and hypertension is observed in In the fully adjusted models, compared with the fourth quintile of birth weight (3.43–3.80 kg), the first quartile (<2.88 kg) was associated with a 25% higher risk of hypertension  .p for interaction = 0.004).In the stratified analysis , the ass2), participants with low birth weight and adult obesity (birth weight <2.88 kg and BMI ≥30 kg/m2) had the highest risk of hypertension . A similar pattern of correlations was observed between birth weight, WHR, and hypertension  and second follow-up , which were 0.82 and 0.81, respectively, indicating a good reliability of self-reported birth weight in the UK biobank. In addition, a previous study of the UK Biobank estimated the associations of self-reported birth weight with sex, deprivation index, multiple births, and maternal smoking, and supported the validity of self-reported birth weight (Our study has several strengths, including a large sample size, long-term follow-up, and strict adjustment for potential confounders. We also estimated the joint effects of birth weight and adult obesity. Several limitations need to be acknowledged . First, h weight . Some oth weight , 40. SecThe current study indicates a nonlinear inverse association between birth weight and hypertension and suggests birth weight between 3.43–3.80 kg as a threshold for reducing disease risk. Low birth weight may interact with adult obesity to increase hypertension risk in adult life. Further studies are warranted to explore underlying mechanisms.The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.DH, AG, and WZ contributed to the conception and design of the study. YZ and DH have full access to all the data in the study and take responsibility for the integrity of the data and the accuracy of the data analysis. YZ, JL, and QL did the statistical analysis and drafted the manuscript. XF and CX critically revised the manuscript for important intellectual content. All authors reviewed and approved the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Sufficient choline and betaine during pregnancy are needed for fetal growth and development.We aimed to investigate the associations between maternal plasma choline and betaine in the third trimester of pregnancy and child growth from birth up to 8 years of age.z scores for weight and height at 1 and 2 y were used as continuous outcome variables. BMI z scores at 1 and 2 y were used as continuous and dichotomous outcomes, and BMI z scores at age 6–8 y were used to study overweight at that age.Concentrations of choline and betaine were measured in plasma of 1331 pregnant women from the KOALA  Birth Cohort Study in the Netherlands. Child weight and height were measured at birth and at 1 (91% complete), 2 (86%), and 6–8 y (76%). Birth weight, weight gain in the first year, and z score  and slightly higher odds of BMI z score >85th percentile  at 1–2 y. Each 1-µmol/L increase of plasma betaine was associated with a mean 12-g  higher weight gain in the first year of life and higher odds of BMI z score >85th percentile at 1–2 y . Lastly, betaine was associated with overweight at 6–8 y , only in boys.Each 1-µmol/L increase of maternal plasma choline was associated with a mean 20-g  higher weight gain in the first year of life, and a higher BMI Third-trimester pregnancy plasma choline and betaine were positively associated with childhood anthropometric measures. In boys, some of the associations may have persisted up to 8 y of age. Further studies may investigate the validity of maternal plasma choline and betaine concentrations as markers of maternal intake and fetal transfer. Choline is an essential nutrient for humans . It is pIn animal models for maternal obesity and gestational diabetes, choline influenced fetal adiposity, insulin resistance, and placental response to hyperglycemia , 15. SimIn the current study we investigated whether maternal plasma concentrations of both choline and betaine in the third trimester of pregnancy were associated with anthropometric measures of the child at birth and at 1 and 2 y of age, and with overweight at 6–8 y of age.n = 2487 in the conventional recruitment group, n = 506 in the alternative recruitment group). From October 2001, coinciding with the start of the recruitment of the alternative group, the pregnant women were also asked to give informed consent for blood collection. The women and their children were followed up during pregnancy and up to 11 y after delivery. The study was approved by the Ethics Committee of the Maastricht University/University Hospital Maastricht. All parents gave written informed consent.This research was performed within the KOALA  Birth Cohort Study in the Netherlands, whose general study design has been described in detail previously . KOALA in = 1703: n = 1197 from the conventional and n = 506 from the alternative recruitment group). Exclusion criteria for the present study were applied in 2 steps: first, multiple pregnancies and newborns with congenital diseases or diseases affecting growth  were excluded, leading to the “candidate cohort”   .A 10-mL maternal blood sample was collected into EDTA-K+-containing tubes between 34 and 36 weeks of gestation. Collection took place during home visits conducted by a trained study nurse. Within 30 min of collection, blood samples were centrifuged on-site at 3000 rpm for 10 min at room temperature. The plasma was transferred into cryogenic vials, transported at 4°C to the laboratory, and stored at −80°C in the biobank until further analysis.Betaine and free choline were measured in EDTA plasma samples by Acquity Ultra Performance Liquid Chromatography coupled to a MicroMass Quattro Premier XE tandem quadrupole mass spectrometer  as described in Kirsch et al. . BrieflyThe mothers filled out questionnaires at 14 and 34 weeks of pregnancy, yielding data on maternal characteristics such as height and weight before pregnancy, gravidity, age, highest achieved educational level, living region within the Netherlands, and country of origin of mother's and father's parents, as well as about smoking and alcohol consumption, and use of supplements containing folic acid.Birth weight as well as pregnancy duration were reported by the mothers at 2 wk postpartum and were verified against midwives’ records. At ages 1 and 2 y, parents were asked to report the most recent weight and height measured at the child health clinic, and to report the children's age (mo) at the time of these measurements . In 3 qu2. In order to standardize children's weight, height, and BMI for gender and actual age of measurement (which could slightly differ among children), measurements were converted to SD scores (z scores) using data from the Dutch reference population as the standard . z scores for weight and height at the ages of 1 and 2 y were used as continuous outcome variables. BMI z scores at 1 and 2 y were used as both continuous and dichotomous outcomes: BMI z scores were dichotomized as <−1.04 and >1.04  (z score >1.04 (corresponding with the 85th percentile) ; all terms were nonsignificant and hence we proceeded with linear models. Birth weight, weight gain in the first year, and ctively) . Overweicentile) .z scores of child weight, length, and BMI at 1 and 2 y we used generalized estimation equations (GEEs) for repeated measurements with an unstructured correlation structure. Logistic GEEs were used to study the associations with the odds of having a BMI z score <−1.04 or >1.04 at 1–2 y of age and with overweight at age 6–8 y. In all GEE models, the age of the child at measurement (d) of the growth outcome was adjusted for, and time-interactions were tested to evaluate whether the results differed by age of the child at measurement ; none of these interactions were statistically significant at α = 0.05 (2-sided), and hence the results are presented for ages 6, 7, and 8 y combined (and expressed as 6–8 y). For the outcome of overweight based on measurements done at home visits (a single measurement between 6 and 8 y), we used logistic regression analysis, adjusting for the same covariates as the GEE analyses.We used multivariable linear regression to analyze the associations of maternal plasma choline and betaine concentrations with birth weight and weight gain in the first year after birth. To analyze the longitudinal association of maternal plasma choline and betaine concentrations with Supplemental Table 1. We started with a model (step 1) that included child variables  and design variables . In step 2 we added ethnicity of the child (based on the country of origin of their grandparents), living region within the Netherlands, maternal education, gravidity, maternal age at the beginning of pregnancy, maternal height, prepregnancy BMI, and alcohol use and smoking in the third trimester. In step 3 we added gestational diabetes as a covariate. We did not adjust for gestational age at birth and birth weight because they are potential mediators and should therefore not be regarded as confounders. Missing values in maternal height for 17 participants were imputed using information on maternal ethnicity, to ensure that the numbers of subjects were identical in each pair of uni- and multivariable models. All results are expressed as regression coefficients (βs) or ORs and 95% CIs per 1 µmol/L choline or betaine and, in the results tables and in order to aid the interpretation, also as effect sizes for an increment of 1 SD of choline or betaine.Potential confounders were selected based on a priori knowledge. For each of the outcomes, several models were tested with and without additional covariates as summarized in Finally, several sensitivity analyses were performed (step 4), including the exclusion of women delivering prematurely (1% of the women) or those with gestational diabetes (1.5%), and effect modification by child's sex, breastfeeding duration , and folic acid supplementation in pregnancy . EffectTo evaluate selective loss to follow-up, maternal baseline characteristics, birth outcomes, children's anthropometric measures, and choline and betaine concentrations were compared between the participants with complete follow-up and those from the candidate and birth cohorts  or drunk alcohol (19%) in the third trimester of pregnancy. Just over half had higher education .Mean ± SD maternal plasma choline and betaine concentrations were 11.3 ± 2.9 µmol/L and 11.5 ± 4.6 µmol/L, respectively. The mean ± SD birth weight was 3560 ± 477 g. The percentage of children born with a birth weight <2500 g was 1.3%, whereas 3.3% were born with a birth weight ≥4500 g (macrosomia). Less than 3% of the children were not of European descent. As We report the results of the fully adjusted models (P = 0.08), with an effect size of 8 g lower birth weight per 1 µmol/L of plasma choline. Plasma betaine was not associated with birth weight .At birth, maternal third-trimester plasma choline showed a tendency toward a negative association with birth weight .Maternal plasma choline and betaine were found to be positively associated with weight gain in the first year of life  (z score > 1.04 (85th percentile) at 1–2 y , but not with the odds of having a BMI z score <−1.04 (15th percentile)  . In addicentile) .z score >1.04 based on parent-reported height and weight)  . These rPinteraction = 0.03) in the models for overweight between ages 6 and 8 y using home-visits data. Stratified analyses revealed a higher OR of becoming overweight between ages 6 and 8 y in boys  but not in girls  per 1 µmol/L of betaine.Performing analyses without year and season of blood collection, adding the covariable breastfeeding duration, or excluding preterm deliveries or women with diabetes in pregnancy yielded no substantially different results. No statistically significant interactions with breastfeeding duration or folic acid supplementation were found either. We found a statistically significant interaction between betaine and child sex (z scores and slightly higher odds of having a BMI z score >1.04 (OR: 1.08). Each 1-µmol/L increase of maternal plasma betaine was associated with 12 g higher weight gain in the first year of life, as well as with higher odds of having a BMI z score >1.04 at 1 and 2 y (OR: 1.03). Moreover, maternal plasma betaine was associated with higher odds of boys being overweight between ages 6 and 8 y (OR: 1.17) when using home-visits data, but not parent-reported data.We studied maternal plasma choline and betaine in the third trimester of pregnancy and their associations with child anthropometric measurements from birth up to the age of 8 y. As main results, we found that each 1-µmol/L increase of maternal plasma choline was associated with a mean 20-g higher weight gain in the first year of life. At 1–2 y, choline was associated with higher BMI z scores at birth in 2 independent cohorts. However, women with plasma choline in the fifth quintile (mean = 11.7 µmol/L) were older, had higher prepregnancy BMI and gestational weight gain, and more frequent gestational diabetes than women in the lowest quintile (mean = 7.1 µmol/L), suggesting that the associations with the anthropometric measurements of the newborn could be confounded or explained by maternal overweight or insulin resistance. With regard to betaine, our results are in agreement with a previous study that reported no association between betaine and birth weight  normalized homeostasis of placental macronutrients  and reduThe strengths of the present study include the large sample size, the longitudinal design, and the possibility to analyze child repeated anthropometric data up to the age of 8 y. Moreover, we had information available on maternal folic acid supplementation in pregnancy and on breastfeeding duration, which allowed us to explore possible effect modification. This study has some limitations as well. First, plasma samples were collected in the nonfasting state; it is unclear whether this may have randomly influenced the measurements, and hence hampered the possibility to detect existing associations. Second, parent-reported child growth data may be less precise than nurse-reported data, therefore decreasing the chance to find a significant association or interaction in the main group of children where data were reported by the parents. Third, BMI was used as a proxy for overweight; the meaning of BMI and overweight at very early ages is debatable, and we cannot exclude that choline and betaine could show associations with body composition (fat and lean mass) that are not mirrored by BMI . Lastly,z scores at 1–2 y (choline only), and higher odds of a BMI z score >85th percentile at 1–2 y. In addition, betaine was associated with higher odds of overweight in 6- to 8-y-old boys. Although the reported associations could be temporary and disappear at older ages, the persistent association that we found of betaine with BMI in boys needs to be confirmed in independent studies. Future studies may address the validity of maternal plasma choline and betaine concentrations as markers of maternal intake and fetal transfer. Research is also warranted to investigate whether pregnancy complications and related risk factors such as maternal overweight modify the associations of choline and betaine with child growth.Maternal plasma choline and betaine concentrations in the third trimester of pregnancy were associated with a higher weight gain of the child in the first year of age, higher BMI nqab177_Supplemental_FileClick here for additional data file.


The aim of the study was to analyze factors predicting (a) the transition to parenthood for female same‐sex couples in Sweden and (b) which partner is the birth mother for the first and (any) second child.Longitudinal studies in which couples become parents are rare for same‐sex couples in any context, even though these families are increasing. Childbearing in lesbian couples is an interesting case for testing theories linked to family utility maximization and household bargaining, as these couples can often choose who will carry a child.Discrete‐time event history and linear probability models are estimated on Swedish population register data (1995–2016) to analyze couples' transitions to first and second birth and the choice of birth mother.The higher the household income and partners' educational levels, the more likely couples are to become parents. However, within‐couple income gaps are small, and income and education are unrelated to the choice of first‐birth mother. Couples are more likely to have a second child and to switch birth mothers if both are highly educated or the first social mother is highly educated.Factors predicting which couples become parents are similar in same‐sex and different‐sex couples. In same‐sex couples, short‐term within‐couple specialization is of little relevance for who becomes the birth mother. Analyses of the transition to a second birth suggest that long‐term planning matters for who becomes the first‐ and second‐birth mother. In recent decades, families have emerged who previously either faced difficulties forming and/or were not legally recognized. Among them are female same‐sex parent families. Same‐sex parenthood was legally recognized in Sweden in 2003, but even before then, registered partner couples and parents could be identified in population registers. Far from all parents register as partners—and not all couples want to become parents—but research indicates that registered partnership (1995–2008) and marriage (from 2009) has been increasingly common among couples who become parents  become parents, the decision regarding To date, we know of no research on the association between socioeconomic factors and the transition to joint parenthood in lesbian couples. In addition, earlier research on within‐couple specialization is predominantly cross‐sectional, retrospective, and based on either surveys or in‐depth interviews with a small group of individuals. Sociodemographic characteristics may be misrepresented in these data due to systematic interview participation and nonresponse  the probability that female, married, or registered partner same‐sex couples become parents by giving birth to a child and (b) who will be the birth mother in these couples, that is, the person who carries the first and (any) second child.In the following, we first discuss theories and earlier research on which couples become parents (between‐couple factors) and second, the choice of birth mother in these couples (within‐couple factors). Thereafter, the theoretical background is summarized, and hypotheses are formulated before we turn to a description of the data and the empirical models. The results are presented, and conclusions are discussed in the two final sections.Focusing on nonfinancial reasons for having children, same‐sex couples and different‐sex couples tend to state similar reasons for wanting to become parents, such as expecting increased happiness, well‐being, and life fulfillment , resulting in more modest legal differences between marriage and cohabitation in Sweden than in many other contexts.For lesbian couples, the costs of assisted procreation or adoption add to the regular expenditures of having children. Couples may choose to privately arrange home insemination, but if they do, the social mother must adopt the child to be legally recognized as a parent in Sweden, and in order to adopt, she has to be married to the birth mother. Neither adoption  nor marriage is needed if the couple conceives at a certified clinic, yet the mothers need to cohabit. The major legal differences between marriage and cohabitation have to do with inheritance rights and the way property is divided in case of a divorce . Some choose to use clinics abroad even for their first child. Couples may go to Denmark to conceive as a way to reduce the wait, obtain more detailed information about the donor, or be able to choose an anonymous donor (the latter is not an option in Sweden). Couples who use clinics abroad pay for the full treatment, making this solution more expensive than using a Swedish clinic. Worth keeping in mind is that becoming parents and having a child is a well‐thought‐out, planned choice for lesbian couples. No couples become pregnant by accident. Although a comparison to different‐sex couples is not the focus of this study, this most likely contributes to a stronger link between income and parenthood in same‐sex compared with different‐sex couples. Earlier research shows that the birth mother in a same‐sex couple on average has a higher income than the mother in a different‐sex couple .In the empirical analyses, the transition to parenthood is seen as a two‐step process where the decision to become parents is made before the couple decides who will be the birth mother. Hence, which married (or registered partner) couples become parents is empirically analyzed first. Based on the literature review, the expectation is that the partner with the lower income and/or education is most likely to be the birth mother (see Table the more resourceful partner will be the (first) birth mother ) provided by Statistics Sweden are analyzed. Data cover the entire population and is available annually until 2016. A total of 4,251 female couples who married in the period 1995 to 2016 were identified. A subsample of these (see below for exclusions) were followed annually from the year of marriage until they had a first child and a second child, if any. The first analysis, the parenthood analysis (P1), follows couples from registered partnership or marriage and onwards to analyze which couples become parents. The second analysis, the first birth mother analysis (M1), analyzes which partner in these couples is the birth mother of the couple's first child. Age is used to discriminate between the two partners, and the younger partner is chosen as the index person. In the third analysis, the second child and parenthood analysis (P2), we follow the parental couples from first birth onward and analyze which ones have a second child. Finally, the fourth analysis, the second birth mother analysis (M2), estimates the likelihood that both children in two‐child families have the same birth mother.Swedish population register data  analysis. The M2 analysis is run on a sample of 510 two‐child families.SEs are clustered at the couple level.A discrete‐time event history model is used for the analysis of which couples become parents where the event is the birth of the couple's first (P1) child and second (P2) child. Episodes are censored in 2016 (the last year of observation) or at the time of divorce, emigration, or death of any of the partners. As the dataset consists of couple‐years, Linear probability models (LPMs) are estimated to analyze which partner is the birth mother in couples who become parents (M1) and have a second child (M2). The analyses are cross‐sectional and estimated in the year of birth of the couple's first and second child, respectively. The M1 outcome is a binary variable indicating that the younger partner (the index person) is the birth mother of the child, and the M2 analysis indicates whether the birth mother of the first child is also the birth mother of the second child.Marriage cohort is included in the P1 and P2 analyses. In P1, it is divided into the categories 1995–2002, 2003–2004, 2005–2008, and 2009–2016. The cut‐off points correspond to the first year that same‐sex partners could adopt (2003), get assisted procreation at a Swedish clinic (2005), and marry (2009). In P2, the two first categories are collapsed because of the small number of couples who had a second child these years (20 in 1995–2002 and 37 in 2003–2004). The partners may have had children before they entered the current marriage. Hence, dichotomous variables indicate whether any of the partners have a previous biological child. These variables  are not included in the P2 and M2 analysis, as very few of the parents who have a second child have children from before .Duration of marriage is included in the P1 and M1 analyses and acts as the time variable in the discrete‐time event history analysis of first parenthood. It is divided into the categories 0–1, 2–3, and >3 years of marriage. The P2 and M2 analyses instead include years since first child birth. It is divided into the categories 0–1, 2, 3, 4, and >4 years in P2, and time varying. In M2, years since first child birth is measured in the year of birth of the second child, and because of small cell sizes, the variable is divided into the categories 0–2 years, 3, and >3 years. Age of the younger partner  and of the first birth mother  in the year of marriage is categorized into 18–30, 31–35, and >35 years of age. The analyses also include the age difference; in the P1 and M1 analyses, whether the older partner is at least 4 years older than the younger partner, and in the P2 and M2 analyses, whether the first birth mother is at least 4 years younger or 4 years older than the first social mother.household income (log)). The logarithmic transformation reduces the likelihood that influential outliers bias the results. Household income is time‐varying and lagged by 1 year in the P1 analysis and measured the year before birth in the P2 analysis. The M1 analysis includes the younger partner's income and her relative income (her income as a share of the household income) the year before birth. The measure is relevant for estimating short‐term family utility and partners' specialization in paid and unpaid work. For those who have their child early in the year, the income the year before may be influenced by pregnancy‐related absences from work. Therefore, additional analyses (available on request) were estimated with income measured 2 years before birth, with similar results as those presented later. The M2 analysis includes the income and relative income of the first birth mother the year before first birth in order not to let the use of parental leave influence the two measures.The P1 and P2 analyses include the logarithm of the sum of the partners' annual gross work incomes .Figure Most couples had their first child early in the marriage. The average time span between marriage and childbirth was 1.5 years Table . Figure n < 20).The partners in the couple were equally likely to be the birth mother of the first child if the child was born in the first year of marriage Figure . From thFigure Table In the parental couples where only one partner was highly educated, the first birth mother was somewhat more likely than the social mother to be the one with higher education.Childbearing in female‐same sex couples was more common among couples and partners who had not previously had a child, and the first birth mother was slightly less likely than the social mother to have had a previous child. Among the parental couples, 38% had a second child during the studied period. The same partner gave birth to both children in 55% of the couples who had a second child Table . RegardiIn the following, the results of the event‐history analyses and LPMs are presented. Cell sizes for categorical variables used in these analyses can be found in The results of the discrete time, event history model of parenthood (P1), and the LPM of first birth mother (M1) are displayed in Table As expected, couples were less likely to have a child the older the younger partner was. The age difference was, for most, not decisive for their likelihood of becoming parents, although there is a negative association (significant at the 10 % level). Childbearing was positively associated with household resources; if one or both partners were highly educated, they were more likely to have a child than if neither was highly educated, and the probability that they became parents increased with household income.The choice of birth mother in the couples who become parents is analyzed in an LPM (M1). The probability that the younger partner carried the child increased with her age (implying that the likelihood that the older partner gave birth decreased with her age). If there was a large age difference in the couple (at least 4 years), the younger partner was more likely than the older partner to be the first birth mother.The choice of the younger partner as the birth mother was positively associated with the older partner having a child previously, but unrelated to the younger partner's previous biological parenthood, all else being equal.Partly because the M1 analysis focuses on couples who, on average, were more highly educated than those who did not transition to parenthood (cf. P1), education was unrelated to the choice of birth mother, as was the younger partner's income and relative income the year before childbirth. Hence, the likelihood that a couple had a child was related to income and educational resources, whereas the choice of birth mother for this child was not.Table The main independent variables in Table The final analysis in Table Returning to the theoretical expectations and hypotheses, the first hypothesis received support, as couples with higher educational and financial resources were more likely to transition to parenthood. The second and third hypotheses, linked to within‐couple short‐term  and long‐term financial reasoning (relative resources/bargaining), did not receive any clear support. Within‐couple resource differences were small in the couples who transitioned to parenthood, which may explain why factors other than income and education matter for who becomes the birth mother. Nevertheless, among those with two children, taking turns giving birth was more common for couples in their upper thirties and in couples where the social mother or both parents were more educated. This may provide tentative evidence for the long‐term family utility perspective  second child. Earlier research .Among the couples who (decide to) become parents, the choice of birth mother may theoretically be linked to short‐term or long‐term family utility. The short‐term perspective predicts a within‐couple division of reproductive behavior and paid work that is in line with Becker's theory on specialization. In other words, the lowest earning partner should be the birth mother for the first and any second child (Hypothesis 2). The long‐term family utility perspective predicts that the higher earning, or more educated partner, is the one who is chosen as the first birth mother as she has the better position in the labor market and thereby can “afford” to be pregnant and take leave (Hypothesis 3). If the parents go on to have a second child, the less resourceful partner will have had time to accumulate more work experience, and the long‐term family utility perspective is strengthened if she carries the second child.In the empirical models, higher educated couples are more likely to transition to a first birth (Model P1) but a partner's income and higher level of education is unrelated to her likelihood of bearing the couple's first child (M1). However, if a couple has a second child, the parents are more likely to switch birth mothers if the first child's social mother or both mothers are highly educated. Hence, the long‐term family utility perspective may influence the decision of who will be the birth mother in families where career‐related costs of childbirth are higher. Within‐couple income differences are small in lesbian couples. Still, opportunity costs of childbirth and leave‐taking differ by educational level, and birth mothers often take the first and the longest parental leave . This research has also received funding from the European Research Council (ERC) under the European Union's Horizon 2020 research and innovation programme awarded to Marie Evertsson (Grant Agreement No. 771770).AppendixS1. Supporting Information.Click here for additional data file.


Are maternal sexually transmitted infections  associated with preterm birth?In this population-based cohort study using US nationwide birth certificate data and including more than 14 million mother-infant pairs with singleton live births, maternal sexually transmitted infections were associated with an increased risk of preterm birth, especially moderately and very preterm birth.These results suggest that maternal sexually transmitted infections may potentially affect neonatal outcomes. This cohort study examines US nationwide birth certificate data for associations between 4 common sexually transmitted infections and preterm birth. Maternal infection has been implicated in the pathogenesis of preterm birth through intrauterine inflammatory response. Chlamydia, gonorrhea, and syphilis are among the most common sexually transmitted infections worldwide, but studies on their association with preterm birth are sparse.To examine the association between maternal chlamydia, gonorrhea, and syphilis infections in pregnancy and the risk of preterm birth in a large population-based study in the US.This population-based retrospective cohort study examined nationwide birth certificate data from the US National Vital Statistics System between 2016 and 2019. All mothers who had a singleton live birth and available data on chlamydia, gonorrhea, or syphilis infection before or during pregnancy and gestational age at birth were included in analysis.Sexually transmitted infection  occurring before or during pregnancy.Preterm birth, defined as gestational age less than 37 weeks.This study included 14 373 023 mothers . Among the mothers, 267 260 (1.9%) had chlamydia, 43 147 (0.3%) had gonorrhea, and 16 321 (0.1%) had syphilis. Among the newborns, 1 146 800 (8.0%) were preterm births. The rate of preterm birth was 9.9%, 12.2%, and 13.3% among women with chlamydia, gonorrhea, and syphilis infection, respectively. After adjustment for sociodemographic and medical and/or health factors, the adjusted odds ratio of preterm birth was 1.03  for chlamydia, 1.11  for gonorrhea, 1.17  for syphilis, and 1.06  for any of these sexually transmitted infections comparing mothers with these conditions and those without.Maternal sexually transmitted infections  were associated with an increased risk of preterm birth. Pregnant women with sexually transmitted infections before or during pregnancy might benefit from targeted prevention for preterm birth. In parallel with the upward trends in preterm birth, the surveillance reports from the US Centers for Disease Control and Prevention (CDC) showed that the prevalence of chlamydia, gonorrhea, and syphilis have been increasing nationally in the general population from 2013 through 2018 in the US.7Maternal infection has been implicated in the pathogenesis of preterm birth through intrauterine inflammatory response.8 However, studies on the association between maternal STIs and preterm birth are sparse, and the results have been inconclusive.11 Inference from some of those previous studies is challenging because they may not have sufficient statistical power because of the relatively low incidence of maternal STIs. The inconsistent findings from previous studies might also be affected by the differences between the study populations. Studies with data from a large, diverse population and available information on potential confounding factors such as detailed sociodemographic information, prenatal factors, and medical and/or health factors are critical to clarifying how maternal STIs may factor into preterm birth.Maternal STIs are known to have adverse effects on the developing fetus.In the US, chlamydia, gonorrhea, and syphilis are routinely screened during pregnancy and documented in birth certificates. Therefore, in this retrospective cohort study, we took advantage of nationwide birth certificate data to examine the association of maternal chlamydia, gonorrhea, and syphilis infections with the risk of preterm birth in more than 14 million mother-infant pairs.13 In the Mother’s Worksheet, maternal characteristics are obtained by hospital staff from the mother during birth registration. For the Facility Worksheet, data are obtained directly from the medical records and include medical and health information.15Federal and state laws require birth certificates to be completed for all births in the US. The NVSS, conducted by the National Center for Health Statistics (NCHS) at the CDC, collects and publishes nationwide data on births from birth certificates according to federal law. Data on live births in all 50 US states and the District of Columbia were collected via the Mother’s Worksheet and the Facility Worksheet.STROBE) reporting guideline for cohort studies.In this population-based study, we used birth data from January 1, 2016, to December 31, 2019, including all 14 373 023 mothers who had a live singleton birth and available data on maternal STIs and gestational age at birth. We used data for these years because the 2003 revised version of the standard birth certificate for live births, which contains more information about sociodemographic and health information, was fully implemented nationwide in 2016 and thereafter. This study was approved as exempt from review and informed consent requirements by the institutional review board at the University of Iowa because it used deidentified data. The conduct and reporting of this study followed the Strengthening the Reporting of Observational Studies in Epidemiology  was collected directly from the medical record by a birth information specialist or clinician responsible for completing the medical and health information for the certificate of live birth using the facility worksheet. According to the NVSS Facility Worksheets Guidebook, a maternal infection was documented if it was present “at the time of the pregnancy diagnosis or a confirmed diagnosis during the pregnancy with or without documentation of treatment.”16 Preterm birth was defined as gestational age less than 37 weeks. We further subdivided the outcome based on gestational age: extremely preterm birth (<28 weeks of gestation), very preterm birth (28-31 weeks of gestation), and moderately preterm birth (32-36 weeks of gestation).2For the ascertainment of preterm birth, gestational age was calculated using the obstetric estimate of gestation at delivery, which was defined as the best obstetric estimate of the infant’s gestation in completed weeks based on the birth attendant’s final estimate of gestation.Information on age, race and ethnicity, education, marital status, height, prepregnancy weight, and smoking during pregnancy were collected using the standardized Mother’s Worksheet during birth registration. Information on parity, previous history of preterm birth, prepregnancy diabetes, prepregnancy hypertension, time of initiation of prenatal care, gestational diabetes, gestational hypertension or preeclampsia, eclampsia, insurance type, and infant sex status was taken from the facility worksheet, which is obtained directly from the medical records.2 tests. We compared the risk of preterm birth for mothers with a specific STI  with those without this specific STI. We also calculated the risk of preterm birth for mothers with any of the 3 included STIs compared with mothers no STIs.We assessed the distribution of sociodemographic and maternal characteristics according to STIs and preterm birth, respectively. Comparisons between categorical variables were tested using χ2R for model 2 was calculated on the basis of multivariable logistic regression to assess the model fit assessment.Multivariable logistic regression models were used to estimate the odds ratios (ORs) and 95% CIs of preterm birth. Multinomial logistic regression models were used to calculate ORs and 95% CIs for the subtypes of preterm birth . In model 1, we adjusted for maternal age , and race and ethnicity . In model 2, we further adjusted for maternal education level , marital status , parity , previous history of preterm birth , prepregnancy body mass index  , prepregnancy diabetes , prepregnancy hypertension , smoking status during pregnancy , timing of initiation of prenatal care , insurance type , and infant sex . These variables were selected based on univariate analysis, biological plausibility and review of the literature. P for interaction was calculated on the basis of multivariable logistic regression models with multiplicative interaction terms (age or race and ethnicity × specific STI or any STIs). We conducted sensitivity analyses by excluding pregnant women who did not receive prenatal care or did not have insurance or medical assistance because these women may not have been screened for maternal STIs during pregnancy. In addition, because a woman may become pregnant more than 1 time between 2016 and 2019, we performed a sensitivity analysis using data from each year. We also conducted another sensitivity analysis, which additionally adjusted for pregnancy complications based on model 2. Pregnancy complications were defined as having gestational diabetes, gestational hypertension, preeclampsia, or eclampsia.We conducted stratified analyses by age, race, and ethnicity to explore whether the association between STIs and preterm birth differed by these demographic factors. P < .0125  were considered statistically significant. All statistical analyses were conducted using survey modules of SAS software version 9.4 (SAS Institute).Two-sided This study included 14 373 023 mothers who delivered singleton live births. Their mean (SD) age was 29 (5.8) years old; 3 435 333 mothers [23.9%] identified as Hispanic, 912 425 [6.4%] as non-Hispanic Asian, 2 058 006 [14.3%] as non-Hispanic Black, and 7 386 568 [51.4%] as non-Hispanic White . Among tIn the overall population, mothers who were infected with chlamydia, gonorrhea, or syphilis had an increased risk of preterm birth compared with mothers without those infections. The adjusted OR of preterm birth was 1.03  for chlamydia, 1.11  for gonorrhea, 1.17  for syphilis, and 1.06  for any STI after adjustment for age, race and ethnicity, education, marital status, parity, previous history of preterm birth, prepregnancy body mass index, prepregnancy diabetes, prepregnancy hypertension, smoking during pregnancy, initiation of prenatal care, insurance type, other infections during pregnancy, and infant sex. We further subdivided preterm birth into moderately, very, and extremely preterm birth. We found a significant association of all 3 STIs examined with moderately preterm birth. The adjusted OR of moderately preterm birth was 1.04  for chlamydia, 1.10  for gonorrhea, and 1.17  for syphilis. Significant associations of gonorrhea and syphilis with very preterm birth were also found. The adjusted OR of very preterm birth was 1.27  for gonorrhea and 1.35  for syphilis .In the stratified analyses by age, race, and ethnicity, significant associations with preterm birth were observed for STI in most age groups. The adjusted OR for any STI was 1.06 , 1.08 , and 1.09  among mothers younger than age 25, between ages 25 and 34, and 35 years or older, respectively. We also found STIs increased the risk of preterm birth among Hispanic , non-Hispanic Black , and non-Hispanic White  mothers. The results for non-Hispanic Asian mothers were less stable, possibly because of the small sample size .The results were robust in the sensitivity analyses that excluded pregnant women who did not receive prenatal care (eTable 2 in the With data from the nationwide population (over 14 million women), we found significant associations between chlamydia, gonorrhea, and syphilis and preterm birth, especially moderately or very preterm birth. The associations were relatively consistent across age and race and ethnicity groups, and the results were robust in a series of sensitivity analyses.19 gonorrhea,18 and syphilis20 with preterm birth. A study from California for chlamydia9 and another study from Washington for gonorrhea21 did not find any significant association with preterm birth, which may be due to the limited sample size in those studies.Previous studies about maternal STIs and preterm birth are sparse, and their findings have been inconsistent. Because of the large sample size in this population-based study, we were able to quantify the association with sufficient statistical power. Consistent with our findings, some, although not all, previous studies showed a significant association of chlamydia,22 Syphilis, which is spread by hematogenous dissemination, may cause systemic infection and placental inflammatory response.22 All of these responses may cause inflammation and activate the maternal and/or fetal immune system, which is an established cause of preterm birth.23 However, whether a maternal infection induces preterm birth may also depend on the characteristic and concentration of the pathogen and the timing of infection.24The potential mechanisms between maternal STIs and preterm birth remain to be elucidated. A potential shared mechanism is inflammation. Chlamydia and gonorrhea, like some other lower genital infections, may ascend via the vagina-cervix and contribute to the occurrence of chorioamnionitis by partial activation of the systemic cytokine network.7 The present study identified maternal STIs  as a novel risk factor for preterm birth, suggesting that addressing maternal STIs might be an unrecognized, preventive approach to reduce preterm birth. The findings support the recommendation by the CDC to screen and treat chlamydia, gonorrhea, and syphilis among pregnant women.This study has important clinical and public health implications. Preterm birth is not only highly prevalent, affecting about one tenth of all births, but also clinically relevant as the leading cause of infant mortality. Therefore, it is imperative to identify modifiable and preventable risk factors for preterm birth. In recent years, there has been an increasing incidence of maternal STIs.The strengths of the current study include the use of the extremely large nationwide data set, which enabled us to estimate the risk of preterm birth in subpopulations and explore the effects of these relatively low incidence diseases on preterm birth. To the best of our knowledge, this is the first study regarding 3 STIs in relation to preterm birth in a large and diverse US population.There are several limitations in our study. First, NVSS does not provide detailed information about treatment for these STIs. Therefore, we cannot further examine the effect modification by treatment on the association between maternal STIs and preterm birth. Second, NVSS did not specify subtypes  for each case of preterm birth. Future studies are needed to assess the association between maternal STIs and preterm birth in these clinical subtypes. Third, although we have adjusted for many factors, we cannot rule out the possibility of potential residual confounding caused by unmeasured or unknown factors. Fourth, there might be misclassification bias for maternal STIs, especially chlamydia and gonorrhea infections, because universal screening in pregnancy for these STIs is recommended but not required. Also, there was no information available to validate the diagnosis and reporting of STIs, which may introduce information bias. While the CDC and American College of Obstetricians and Gynecologists have published recommendations for screening of pregnant women for STIs, it is possible that not all clinics in the US have followed those recommendations stringently. Fifth, we don’t know the exact timing of onset for these STIs. However, we adjusted the initiation of prenatal care in our model because serologic tests for these STIs should be performed for all pregnant women and high-risk women at their first prenatal visit according to the Sexually Transmitted Diseases Treatment Guidelines in the US.In this large national study in the US, maternal STIs of chlamydia, gonorrhea, and syphilis were associated with an increased risk of preterm birth, especially of moderately and very preterm birth. These findings suggest that pregnant women with STIs before or during pregnancy warrant targeted prevention for preterm birth. Future investigation is needed to understand the underlying mechanisms.