**The relationship between gaze stability, retinal image quality, and visual perception is complex. Gaze instability related to pathology in adults can cause a reduction in visual acuity (e.g., Chung, LaFrance, & Bedell, 2011). Conversely, poor retinal image quality and spatial vision may be a contributing factor to gaze instability (e.g., Ukwade & Bedell, 1993). Though much is known about the immaturities in spatial vision of human infants, little is currently understood about their gaze stability. To characterize the gaze stability of young infants, adult participants and 4- to 10-week-old infants were shown a dynamic random-noise stimulus for 30-s intervals while their eye positions were recorded binocularly. After removing adultlike saccades, we used 5-s epochs of stable intersaccade gaze to estimate bivariate contour ellipse area and standard deviations of vergence. The geometric means (with standard deviations) for infants' bivariate contour ellipse area were left eye = −0.697 ± 0.534 log(° ^{2}), right eye = −0.471 ± 0.367 log(°^{2}). For binocular vergence stability, the infant geometric means (with standard deviations) were horizontal = −1.057 ± 0.743 log(°), vertical = −1.257 ± 0.573 log(°). These values were all not significantly different from those of the adult comparison sample, suggesting that gaze instability is not a significant limiting factor in retinal image quality and spatial vision during early postnatal development.**

^{2}; for black pixels it was 11 cd/m

^{2}. A field of randomly placed green noise elements was drawn on a black background using routines from the Psychophysics Toolbox (Brainard, 1997) and MATLAB (MathWorks, Natick, MA). The noise elements each subtended visual angles of 4° horizontally and 6° vertically. The location of each element was assigned using a uniform random-number generator and could overlap other nearby noise elements. Between 45% and 50% of the screen was green in each frame. The stimulus then refreshed to a new field of static noise elements at 3 Hz.

*χ*

^{2}represents the chi-square value corresponding to a probability of 0.95,

*S*

_{H}and

*S*

_{V}are the standard deviations of monocular horizontal and vertical gaze position, and

*ρ*is the Pearson product–moment correlation coefficient between horizontal and vertical gaze position. The BCEA value represents the area of an ellipse drawn around the central 95% of the fitted distribution of eye positions in two dimensions, and thus smaller values mean less variation in eye position (although see Appendix B). Though a number of epochs of stable fixation could occur within an individual's visit, only the epoch with the median BCEA was used for parametric summary analyses.

*t*test was performed to test the equality of the adult and infant data (

*α*= 0.05).

*SD*= −0.697 ± 0.534 log(°

^{2}), mean equivalent to 0.470°

^{2}; right: −0.471 ± 0.367 log(°

^{2}), mean equivalent to 0.475°

^{2}) and adults (left: −0.450 ± 0.333 log(°

^{2}), mean equivalent to 0.440°

^{2}; right: −0.522 ± 0.383 log(°

^{2}), mean equivalent to 0.430°

^{2}) yielded no statistically significant difference for either eye—left:

*t*= 1.328,

*p*= 0.203, 95% confidence interval for the geometric mean difference [−0.146, 0.639]; right:

*t*= 0.331,

*p*= 0.744, 95% confidence interval [−0.051, 0.153]. The group standard deviations were also similar—left: infants = 0.534 log(°

^{2}), adults = 0.333 log(°

^{2}); right: infants = 0.367 log(°

^{2}), adults = 0.383 log(°

^{2}).

*SD*) for the two subjects were −0.723 ± 0.08 log(°

^{2}), mean equivalent to 0.189°

^{2}, and −0.790 ± 0.08 log(°

^{2}), mean equivalent to 0.162°

^{2}. These mean values are plotted on the right-hand side of Figure 3.

*SD*) for horizontal vergence for infants was −1.057 ± 0.743 log(°), mean equivalent to 0.087°, and for adults it was −1.156 ± 0.416 log(°), mean equivalent to 0.069°. The corresponding vertical geometric mean (±

*SD*) for infants was −1.257 ± 0.573 log(°), mean equivalent to 0.055°, and for adults it was −1.311 ± 0.521 log(°), mean equivalent to 0.049°. Neither horizontal (

*t*= 0.405,

*p*= 0.691) nor vertical (

*t*= 0.239,

*p*= 0.813) geometric means were significantly different between infants and adults.

*t*= 1.27,

*p*= 0.216).

^{2}) when assessing children (5–17 years) over 30-s intervals with a microperimeter. Gonzalez et al. (2012) measured adults for 15-s intervals using an EyeLink 1000 and found log BCEAs of −0.88 log(°

^{2}) for binocular viewing, similar to the numbers found for two adult subjects during our control experiment, in a shorter interval with the same eye tracker: −0.723, −0.790 log(°

^{2}). Additional data from Cherici, Kuang, Poletti, and Rucci (2012) demonstrate that untrained subjects (like those in these experiments) have considerably larger estimates of fixation instability, especially without a defined fixation target, which may also have inflated the adult gaze instabilities measured here.

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*SD*= ±18 ms), and durations were normally distributed. The average adult saccade lasted 46 ms (

*SD*= ±9 ms). These were not significantly different from one another (

*t*= 1.60,

*p*= 0.113). The median number of saccades in analyzed epochs was seven for infants (interquartile range = 5–9.5) and one for adults (interquartile range = 0–2.75). To examine the possible effect of saccade number and duration on the results, a 5-s epoch of adult data was identified for which no saccades were flagged. Simulated saccades varying in length between 20 and 100 ms were inserted into this vector at random starting points. If the next starting point was within a simulated saccade, the following new starting point was used. The number of inserted saccades varied between 1 and 50; the length of the entire epoch remained constant at 5 s. The standard deviation of the normalized intersaccade intervals was then computed.

*SD*) was 0.181°

^{2}± 0.064, and for infants it was 0.248°

^{2}± 0.107. These distributions were not statistically significantly different as a function of age (

*t*= 1.899,

*p*= 0.071). This additional analysis supported the use of the bivariate contour ellipse area for comparison with other studies in the literature and increased our confidence that the normality of the data did not impact the conclusion.